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On model fitting for multivariate polytomous response data

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On model fitting for multivariate polytomous response data
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Thesis (Ph. D.)--University of Florida, 1992.
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Includes bibliographical references (leaves 193-199).
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by Joseph B. Lang.

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ON MODEL FITTING FOR MULTIVARIATE POLYTOMOUS
RESPONSE DATA










By

JOSEPH B. LANG


A DISSERTATION PRESENTED TO THE GRADUATE SCHOOL
OF THE UNIVERSITY OF FLORIDA IN PARTIAL FULFILLMENT
OF THE REQUIREMENTS FOR THE DEGREE OF
DOCTOR OF PHILOSOPHY

UNIVERSITY OF FLORIDA

1992


UNIVERSITYY OF FLORIDA US1llic15













ACKNOWLEDGMENTS


I would like to express my appreciation to Dr. Alan Agresti for serving

as my dissertation advisor. For the many comments, ideas, and lessons he

has shared with me, I am greatly indebted. Through his advisement and

guidance, he has taught me to appreciate and respect good statistical research

and teaching. He is a mentor worthy of emulation. I also want to express

my gratitude to Dr. Jane Pendergast, who also served on my dissertation

committee. I learned a great deal from her during the two years that I worked

in the Biostatistics Department. To all of the faculty at the University of

Florida, I extend my thanks. The statistics department, with its scholarly

and friendly atmosphere, proved to be a wonderful place to learn.

The influences of persons from my past are not forgotten. Without

Patrick Kearin's stimulating teaching of high school math, I may never have

become interested in this subject. The genuine excitement delivered by Dr.

James Kepner, in his teaching of undergraduate statistics, was the reason I

decided to pursue an advanced degree in statistics.

I would like to thank my parents and the rest of my family for all of the

support and encouragement they have given over the course of my studies

and research. My friends and student colleagues deserve many thanks as

well. Finally, I would like to thank Kendra Paar for always being there to

support and encourage me while I was writing this paper.













TABLE OF CONTENTS


page
ACKNOWLEDGMENTS ............................................ ii

LIST OF TABLES ................................................ v

ABSTRACT .................................................... ... vi

CHAPTERS

1 INTRODUCTION ............................................. 1

1.1 A Brief Introduction to the Problem...................... 1
1.2 Outline of Existing Methodologies-No Missing Data ...... 3
1.3 Outline of Existing Methodologies-Missing Data.......... 12
1.4 Format of Dissertation ...................................... 14

2 RESTRICTED MAXIMUM LIKELIHOOD FOR A
GENERAL CLASS OF MODELS FOR
POLYTOMOUS RESPONSE DATA .................... 17

2.1 Introduction ........................................ .. ..... 17
2.2 Parametric Modeling-An Overview....................... 24
2.2.1 Model Specification .................................. 25
2.2.2 Measuring Model Goodness of Fit ................... 33
2.3 Multivariate Polytomous Response Model Fitting .......... 43
2.3.1 A General Multinomial Response Model.............. 44
2.3.2 Maximum Likelihood Estimation .................... 48
2.3.3 Asymptotic Distribution of Product-Multinomial
M L Estimator ...... ..... ..... ....... .... .......... 56
2.3.4 Lagrange's Method-The Algorithm ................ 60
2.4 Comparison of Product-Multinomial and
Product-Poisson Estimators ........................... 67
2.5 Miscellaneous Results ....................................... 78
2.6 Discussion ................................................... 83










3 SIMULTANEOUSLY MODELING THE JOINT AND
MARGINAL DISTRIBUTIONS OF MULTIVARIATE
POLYTOMOUS RESPONSE VECTORS .................. 87

3.1 Introduction............................................... 87
3.2 Product-Multinomial Sampling Model..................... 88
3.3 Joint and Marginal Models................................. 93
3.4 Numerical Examples ....................................... 98
3.5 Product-Multinomial Versus Product-Poisson
Estimators: An Example .......................... 111
3.6 Well-Defined Models and the Computation of
Residual Degrees of Freedom ......................... 121
3.7 Discussion .............. .................................... 132

4 LOGLINEAR MODEL FITTING WITH
INCOMPLETE DATA...................................... 135

4.1 Introduction .............. ................................... 135
4.2 Review of the EM Algorithm................................137
4.2.1 General Results .................. ....................138
4.2.2 Exponential Family Results ........................... 140
4.3 Loglinear Model Fitting with Incomplete Data............. 144
4.3.1 The EM Algorithm for Poisson Loglinear Models..... 145
4.3.2 Obtaining the Observed Information Matrix ..........148
4.3.3 Inferences for Multinomial Loglinear Models ..........152
4.4 Latent Class Model Fitting-An Application .............. 160
4.5 Modified EM/Newton-Raphson Algorithm................. 166
4.6 Discussion .................................................. 170

APPENDICES

A CALCULATIONS FOR CHAPTER 2.........................172

B CALCULATIONS FOR CHAPTER 4.........................176

BIBLIOGRAPHY ................... ..........................193

BIOGRAPHICAL SKETCH ........................................ 200













LIST OF TABLES


page
2.1 Opinion Poll Data Configuration................................. 22

3.1 Interest in Political Campaigns ................................... 91

3.2 Cross-Over Data............ .......... ..... ....... ................ 92

3.3 Joint Distribution Models-Goodness of Fit..................... 100

3.4 Marginal Distribution Models-Goodness of Fit............... 101

3.5 Candidate Models in J(L x L + D) n M(U)-Goodness of Fit... 102

3.6 Estimates of Freedom Parameters for
Model J(L x L + D) n M(CU)..................................... 103

3.7 Freedom Parameter Estimates and Standard Errors.............. 105

3.8 Estimated Cell Means and Standard Errors ................. 106

3.9 Cross-Over Data Models-Goodness of Fit....................... 110

3.10 Freedom Parameter ML Estimates for Model J(UA) n M(U) .... 110

3.11 Children's Respiratory Illness Data........................... 112

3.12 Product-Multinomial versus Product-Poisson Freedom
Parameter Estimation ........................................ 117

4.1 Observed cross-classification of 216 respondents
with respect to whether the tend toward
universalistic (1) or particularistic (2) values
in four situations (A,B,C,D) of role conflict ................. 162

4.2 Parameter and Standard Error Estimates ....................... 164

4.3 Classification Probability Estimates ..................... ....... 165








Abstract of Dissertation Presented to the Graduate School
of the University of Florida in Partial Fulfillment of the
Requirements for the Degree of Doctor of Philosophy

ON MODEL FITTING FOR MULTIVARIATE POLYTOMOUS
RESPONSE DATA

By

Joseph B. Lang

May, 1992

Chairman: Dr. Alan Agresti
Major Department: Statistics

A broad class of models that imply structure on both the joint

and marginal distributions of multivariate categorical (ordinal or nominal)

responses is introduced. These parsimonious models can be used to si-

multaneously describe the marginal distributions of the responses and the

association structure among the responses. As a special case, this class

of models includes classical log- and logit-linear models. In this sense,

we address model fitting for multivariate polytomous response data from

a very general perspective. Simultaneous models for joint and marginal

distributions are useful in a variety of applications, including longitudinal

studies and studies dealing with social mobility and inter-rater agreement.

We outline a maximum likelihood fitting algorithm that can be used for

fitting a large class of models that includes the class of simultaneous models.

The algorithm uses Lagrange's method of undetermined multipliers and a

modified Newton-Raphson iterative scheme. We also discuss goodness-of-fit

tests and model-based inferences. Inferences for certain model parameters

are shown to be equivalent for product-Poisson and product-multinomial

vi







sampling assumptions. This useful equivalence result generalizes existing

results. The models and fitting method are illustrated for several applications.

Missing data are often a problem for multivariate response data. We

consider inferences about loglinear models for which only certain disjoint

sums of the data are observable. We derive an explicit formula for the

observed information matrix associated with the loglinear parameters that

is intuitively appealing and simple to evaluate. The observed information

matrix can be evaluated at the maximum likelihood estimates and inverted

to obtain an estimate of the precision of the loglinear parameter estimates.

The EM-algorithm can be used to fit these incomplete data loglinear models.

We describe this algorithm in some detail, paying special attention to the

Poisson loglinear model fitting case. Alternative fitting algorithms are also

outlined. One proposed alternative uses both the EM and Newton-Raphson

algorithm, thereby resulting in a faster, more stable, algorithm. We illustrate

the utility of these results using latent class model fitting.












CHAPTER 1
INTRODUCTION


1.1 A Brief Introduction to the Problem

There are many situations when multiple responses are observed for each

'subject' in a group, or several groups. Here 'subject' is generically used to

refer to a randomly chosen object that generates responses. The multiple

responses could represent repeated measurements taken on subjects over time

or occasions. They could be the ratings assigned by several judges that all

viewed and rated the same set of slides (here, the 'subjects' are the slides).

Or, perhaps, it may be that several distinct or noncommensurate responses

are recorded for each subject. These responses are often categorical-ordinal

or nominal-and inevitably interrelated. This dissertation addresses issues

related to modeling and model fitting for multivariate categorical (ordinal or

nominal) responses.

Models for multivariate categorical response data are usually developed

to answer questions about (i) the association structure among the multiple

responses or (ii) the behavior of the marginal distributions of the response

variables. Specifically, a typical question of the first type is, "How are the

responses interrelated and is this interrelationship the same across the levels

of the covariates?" A typical type ii question is, "How do the (marginal)

responses depend on the covariates or occasions?" Historically, many models

(e.g. log- and logit-linear models) have been developed for the primary

-1-






-2-

purpose of answering the type i questions. Many of these models can easily

be fitted using maximum likelihood (ML) methods. These models typically,

however, are not useful for answering the type ii questions (Cox, 1972).

Marginal models-those models used to answer type ii questions-are not

as well developed. One reason for this is that ML fitting of these marginal

models is more difficult. At present, the method of weighted least squares

(WLS) is used almost exclusively for fitting these models.

Suppose that we are interested in answering questions of both types

i and ii. Usually the questions are addressed using two different models, a

joint distribution model and a marginal model, and fitting them separately. It

seems reasonable to want a model that can be used to address simultaneously

both questions. That is, we would like a model that simultaneously implies

structure on both the joint and marginal distribution parameters. To date,

there has been very little work done on the development and fitting of these

simultaneous models.

Whenever multiple responses are observed it is inevitable that there will

be missing data. There are several ways to fit the Poisson loglinear model with

incomplete data. One popular method is to use the EM algorithm to find the

ML estimates of the loglinear parameters. One drawback to this algorithm

is that a precision estimate of the ML estimators is not produced as a by-

product. Several numerical techniques have been developed to approximate

the observed information matrix, which, upon inversion, will act as the

precision estimate. However, it would be of some convenience to derive an

explicit formula for the observed information matrix, at least in some special

cases.






-3-

1.2 Outline of Existing Methodologies-No Missing Data



We begin our discussion by considering the case of no missing data.

There are many methods for analyzing multivariate categorical (ordinal or

nominal) response data. These methods usually involve fitting (separately)

models for the joint or the marginal distributions of the response vectors.

In rare instances, simultaneous models for both the joint and marginal

distributions are considered. Maximum likelihood fitting methods for the

joint distribution models are simple and described in almost every standard

text on categorical data analysis. The fitting of marginal models using

ML methods is more difficult. Maximum likelihood fitting of the marginal

homogeneity model was considered by Madansky (1963) and Lipsitz (1988).

The fitting of a more general class of marginal models was considered

by Haber (1985a). Finally, the fitting of simultaneous models using ML

methods has only been addressed in the bivariate response case. The fitting

technique becomes very complicated when there are more than two categorical

responses. To appreciate the complexity of extending the technique to

multivariate response data, see section 6.5 of McCullagh and Nelder (1989)

or perhaps Dale (1986). In contrast, the ML fitting method of Chapter 2 can

easily be used to fit many marginal and simultaneous models. In the next few

paragraphs, we briefly describe the existing methods for modeling and model

fitting for multivariate categorical response data.

Modeling Joint Distributions Separately. One common method for analyz-

ing multivariate categorical responses is to model the joint distribution only.

These models, which include classical log- and logit-linear models for the






-4-

joint probabilities, are useful for describing the association structure among

the responses. The last 30 years have seen the development of these methods

for analyzing multivariate categorical responses (Haberman, 1979; Bishop et

al., 1975; Agresti, 1984, 1990). For specificity, consider the following panel

study: One hundred randomly selected subjects were asked how interested

they were in the political campaigns. They were to respond on the 3 point

ordinal scale, (1) Not Much, (2) Somewhat, and (3) Very Much. Then four

years later the same group of subjects was asked to respond on the same

scale to the same question. A separate investigation into the association

structure would enable us to answer questions of a conditional nature. For

example, we could estimate the probability of responding 'Very Much' on the

second occasion given that the response at the first occasion was 'Not Much'.

The description of these 'transitional' probabilities, although very interesting,

may not be completely satisfactory. We may also be interested in addressing

questions with regard to the marginal distributions. Perhaps we would like

to answer the question, "Are the distributions of responses to the political

interest question the same for each occasion?" Laird (1991), in a nice review of

likelihood-based methods for longitudinal analysis, mentions that the utility

of classical log- and logit-linear models is restricted to two situations: (1)

modeling the dependence of a univariate response on a set of covariates and

(2) modeling the association structure between a set of multivariate responses.

These models place structure on the joint probabilities and so they are not

directly useful for studying the dependence of the marginal probabilities on

occasion and other covariates. This problem was pointed out by several

authors (Cox, 1972; Prentice, 1988; McCullagh and Nelder, 1989;






-5-

Liang et al., 1991). An advantage of these models is that they are simple to fit

using either WLS (Grizzle et al., 1969), ML (McCullagh and Nelder, 1989),

or iterative proportional fitting (Bishop et al., 1975) methods. There are

many standard statistical programs available for fitting these models (SAS,

SPSS BMDP, GLIM, GENSTAT).

Modeling Marginal Distributions Separately. A second approach to an-

alyzing multivariate categorical responses is to model only the marginal

distributions and to ignore the joint distribution structure. Full likelihood

methods that consider only models for the marginal probabilities tacitly

assume a saturated model for the joint distribution. Therefore, the models

may be far from parsimonious. In the non-Gaussian response setting, there

is a distinction between these marginal models and the transitional (or

conditional) models of the previous paragraph. Marginal models describe the

occasion-specific distributions and the dependence of those distributions on

the covariates. Transitional or conditional models describe the distribution

of individual changes over occasions. Models for these transitions can be

represented as probability distributions for the future state 'given' the past

states. Questions regarding transition probabilities can only be investigated

with longitudinal data. On the other hand, questions regarding the marginal

probabilities could theoretically be answered using cross-sectional data,

provided the cohort (subject) effects were negligible. Panel studies resulting

in longitudinal data result in more powerful tests for significance of within

cluster factors, such as occasion effect. This follows because there is a reduced

cohort effect; we are using the same panel of subjects at each occasion. For






-6-

further discussion about the distinction between marginal and transitional

models, see Ware et al. (1988), Laird (1991), and Zeger (1988).

We will briefly discuss existing methods for making inferences about

the marginal probabilities separately. We will group these methods into 5

categories: (1) nonmodel-based methods, (2) WLS methods, (3) ML methods,

(4) Semi-parametric methods, and (5) other methods.

Nonmodel-based methods can be used to derive test statistics used for

testing specific hypotheses regarding the marginal distributions. Examples

include the Cochran-Mantel-Haenszel (1950, 1959) statistic which can be used

for testing the hypothesis of marginal homogeneity (MH) (cf. White et al.,

1982), McNemar's (1947) statistic which can be used for testing the equality of

two dependent proportions, and Madansky's (1963) likelihood-ratio statistic

for MH. Madansky's statistic is a difference in fit of the model of marginal

homogeneity to the fit of the unstructured (saturated) model (see also Lipsitz,

1988 and Lipsitz et al., 1990). Many other relevant test statistics, some of

which are generalizations or modifications of the aforementioned (cf. Mantel,

1963; White et al., 1982), exist. Cochran's (1950) Q statistic and Darroch's

(1981) Wald-type statistic are examples of other test statistics that can be

used to test for marginal homogeneity.

Presently, if one was to fit a marginal model, say a generalized loglinear

model of the form Clog Ai = X/, where p is the vector of expected counts

in the full contingency table, he or she would most likely use the WLS fitting

algorithm. Most statistical software that fits these generalized loglinear

models does so using WLS. There are some advantages to using WLS. It

is computationally simple. Second-order marginal information is all that is






-7-

needed. And, the estimates are asymptotically equivalent to ML estimates.

Some disadvantages are that covariates must be categorical, sampling zeroes

create problems, and estimates are sensitive when second-order marginal

counts are small. The WLS method for analyzing categorical data was

originally outlined by Grizzle, Starmer and Koch (1969). Subsequently,

marginal models for longitudinal categorical data, or more generally mul-

tivariate categorical response data, have been introduced and fitted using the

WLS method (Koch et al., 1977; Landis and Koch, 1979; Landis et al., 1988;

Agresti, 1989).

Maximum likelihood fitting of marginal models is more difficult since

the model utilizes marginal probabilities, rather than joint probabilities to

which the likelihood refers. When the responses are correlated, as they

invariably are, the marginal counts do not follow a product-multinomial

distribution. The full-table likelihood must be maximized subject to the

constraint that the marginal probabilities satisfy the model. Haber (1985a)

considers fitting generalized loglinear models of the form C log Ap = XP3 using

Lagrange multipliers and an unmodified Newton-Raphson iterative scheme.

The algorithm becomes very difficult to implement for even moderately large

tables. This is primarily due to the difficulty of inverting the large Hessian

matrix of the Lagrangian objective function. In this dissertation we consider a

modified Newton-Raphson that uses a much simpler matrix than the Hessian.

The matrix is easily inverted even for relatively large tables. Haber (1985b)

considers the estimation of the parameters / in the special case Clog y = XP3.

We will use a modification of the method of Aitchison and Silvey (1958, 1960)

and Silvey (1959) to investigate the asymptotic behavior of the estimators of






-8-

3 in the more general model Clog Ap = XP3, thereby extending the work of

Haber (1985b). Another relevant paper, Haber and Brown (1986), considers

ML fitting of a model for the expected counts p that has loglinear and

linear constraints. One can test hypotheses about the marginal probabilities

by comparing the fit of relevant models. Haber (1985a, 1985b) and Haber

and Brown (1986) only consider fitting the marginal models separately. No

attempt has been made to simultaneously model the joint and marginal

distributions.

Semi-parametric methods such as quasi-likelihood (Wedderburn, 1974)

and a multivariate extension, generalized estimating equations (GEE), have

become popular in recent years. The work of Liang and Zeger (1986), which

advocated the use of these GEEs, has been extended to cover the multivariate

categorical response data setting (Prentice, 1988; Zhao and Prentice, 1991;

Stram et al., 1988; Liang et al., 1991). With these semi-parametric methods,

the likelihood is not completely specified. Instead, generalized estimating

equations are chosen so that, when the marginal model holds, even if the

association among the multiple responses is misspecified, the estimators are

consistent and asymptotically normally distributed. These estimators, used

in conjunction with a robust estimator of their covariance (Liang and Zeger,

1986; Zeger and Liang, 1986; White, 1980, 1981, 1982; Royall, 1986), result

in consistent inference about the effects of interest. When the responses are

truly independent, the estimating equations with correlation matrix taken to

be the identity matrix, are equivalent to the likelihood equations. The GEE

approach requires the specification of a 'working' association or correlation

matrix. Examples of working associations include those that imply all






-9-

pairwise associations (measured in terms of odds ratios) are the same and

that the higher order associations are negligible (Liang et al., 1991).

A related approach is known as GEE2. The consistency of these esti-

mators follows only if both the marginal model and the pairwise association

model are correctly specified. This approach is a second order extension

of the GEEs of Liang and Zeger (1986) which are now termed GEE1. It

is second order because the estimation of the marginal model parameters

and the pairwise association model parameters is considered simultaneously.

The focus of both approaches, GEE1 and GEE2, is usually on modeling

the marginal distributions-investigating how the marginal distributions

depend on occasion and covariates. The association is considered a nuisance.

Presently, there are no tests for goodness-of-fit of these models and so the

investigation into how well both models fit can be done only at an empirical

level. The assumption that higher order effects are negligible may not be

tenable. Testing procedures to assess the validity of these assumptions have

yet to be developed. Also, in contrast to WLS and ML methods, which

require only that the missing data be 'missing at random' (MAR), the semi-

parametric approaches require the missing data to be 'missing completely

at random' (MCAR). The assumption that the missing data mechanism is

MCAR is a much stronger assumption than MAR (Little and Rubin, 1986).

Finally, there are many other approaches to analyzing the marginal

probability structure separately. There are random effects models, whereby

subject-specific random effects induce a correlation structure on the multiple

responses. The marginal approach-the full likelihood is obtained by

averaging across the random effects-is computationally difficult (Stiratelli






10-
et al., 1984). An alternative is to condition on the sufficient statistics

for the subject effects and consider finding the estimates by maximizing

the conditional likelihood. For further details on these conditional and

unconditional methods see Rasch, 1961; Tjur, 1982; Agresti, 1991; Stiratelli

et al., 1984; Conaway, 1989, 1990. As yet another alternative, Koch et al.

(1980) give a bibliography for relevant nonparametric methods for analyzing

repeated measures data. Agresti and Pendergast (1986) consider replacing

the actual observations by their within cluster rank and testing for marginal

homogeneity using the ordinary ANOVA statistic for repeated measures data.

A three-stage estimator for repeated measures studies with possibly missing

binary responses has been developed by Lipsitz et al. (1992). This approach

is very similar to a generalized least squares approach, but it has some of

the nice features of the GEE approaches. One of these nice features is that

the estimators and their variance estimates are consistent under very mild

assumptions. An extension of this method to the polytomous response case

has yet to be developed.

Simultaneous Investigation of Joint and Marginal Distributions. There

has been very little work done to investigate simultaneously the joint and

marginal distribution structure. In some ways GEE2 is an attempt to

describe both distributions. However, only the pairwise (not the joint)

association structure is modeled; the higher-order associations are considered

a nuisance. Tests comparing nested models have not been developed in this

semi-parametric setting. Full likelihood approaches have been addressed

by Dale (1986), McCullagh and Nelder (1989, Chapt. 6), and Becker and

Balagtas (1991). Dale models the joint distributions of bivariate ordered






11-

categorical responses by assuming that the log global odds ratios follow a

linear model. The marginal probabilities are assumed to follow a cumulative

logit model. McCullagh and Nelder consider simultaneously modeling the

joint and marginal probabilities of a bivariate dichotomous response (two

distinct responses) by assuming that the log odds-ratios follow a linear

model and that the marginal probabilities follow a logit-linear model. Their

example included age as a categorical covariate. Finally, Becker and Balagtas

consider models for two-period cross-over data. The bivariate dichotomous

response was the response to the two different treatments. Order of treatment

application was considered a covariate. They assumed that the two log odds

ratios followed a linear model and that the marginal probabilities satisfied a

loglinear model. Because it is the marginal probabilities and not the joint

probabilities that satisfy a loglinear model, Becker and Balagtas refer to the

model as log nonlinear.

The ML model fitting approach used by each of these authors involves

a reparameterization of the likelihood, which is a function of the joint

probabilities, in terms of the joint and marginal model parameters. The

reparameterization in the bivariate response case-the case each author

considered-is somewhat complicated especially for multi-level responses. To

make matters worse, the extension of this method to general multivariate

polytomous responses looks to be extremely difficult. If the repaparameter-

izations are made so that the full likelihood is expressible in terms of the

joint and marginal model parameters, the likelihood can be maximized using

a Newton-Raphson-type algorithm. Basically, one must solve for the root of

some nonlinear score equation. This maximization approach is very sensitive






12-

to the starting value in that convergence to a local maximum is not likely

unless the starting estimate is very close to the actual maximum. Finding

reasonable starting values is not a simple task. Dale (1986) outlines a method,

specifically for the models considered in that paper, for finding a starting

estimate.

In this dissertation, we outline an ML fitting method that can easily be

used to fit a large class of simultaneous models, including those considered

by Dale, McCullagh and Nelder, and Becker and Balagtas. The approach

involves using Lagrange's method of undetermined multipliers along with a

modified Newton-Raphson iterative scheme. For all of the models considered,

an initial estimate for the algorithm is the data counts themselves along with

a vector of zeroes corresponding to a first guess at the values of the Lagrange

multipliers. The convergence of the algorithm is quite stable. The extension

to multivariate polytomous response data is straightforward.


1.3 Outline of Existing Methodologies-Missing Data

Missing data is often an issue when the response is multivariate in nature.

Missing data can also occur in more hypothetical situations. Examples

include loglinear latent class models (Goodman, 1974; Haberman, 1988)

and linear mixed or random effects models (Laird et al., 1987). In latent

class analyses, a latent variable, which is unobservable, is assumed to exist.

Mixed or random effects models posit the existence of some unobservable

random variables that affect the mean response. In this brief outline, we will

consider ML methods for model fitting when the data are not completely

observable. Little and Rubin (1986) provide a nice summary of methods






13 -
for model fitting with incomplete data. There are many ways to find the

maximum likelihood estimators when the data are not completely observable,

each method having its positive and negative features. We could work directly

with the incomplete-data likelihood, which is usually complicated relative to

the complete-data likelihood, and use a Newton-Raphson or Fisher-scoring

algorithm. Palmgren and Ekholm (1987) and Haberman (1988) use these

methods to obtain maximum likelihood estimates and their standard errors.

Alternatively, we could avoid the complicated likelihood altogether and use

the Expectation-Maximization algorithm (Dempster et al., 1977). Sundberg

(1976) discusses the properties of the EM algorithm when it is used to

fit models to data coming from the regular exponential family. The EM

algorithm is one of the more flexible ML fitting algorithms for missing data

situations. We will primarily focus on this method for fitting loglinear models

with incomplete data.

Although the EM algorithm is easily implemented to fit loglinear models

with incomplete data, the algorithm does not provide an estimate of precision

of the model parameter estimators. Meng and Rubin (1991) outline a

supplemental EM (SEM) algorithm, whereby, upon convergence of the EM

algorithm, the variance matrix for the model estimators is adjusted to account

for missing data. The adjustment is a function of the rate of convergence of

the EM algorithm, which in turn is a function of how much information

is missing. Meng and Rubin numerically estimate the rate of convergence,

thereby obtaining an estimate of precision that reflects missingness. Although

this approach should prove to be applicable in the general situation, it still

is desirable to derive an explicit formula for the variance matrix that reflects






14-

missingness. Other authors (Meilijson, 1989; Louis, 1982) have discussed

methods for estimating precision of model estimators when the data are

incomplete and the EM algorithm is used. Meilijson's method involves EM-

aided differentiation, which is essentially a numerical differentiation of the

score vector. The method relies on the assumption that the observed data

components are i.i.d. (identically and independently distributed). Louis

gives an analytic formula for the observed information matrix based on the

incomplete data. The computation of the observed information matrix based

on this formula is not straightforward and must be considered separately for

each special application.


1.4 Format of Dissertation

In Chapter 2, we develop a maximum likelihood method for fitting a large

class of models for multivariate categorical response data. This development

follows a general discussion about parametric modeling. Concepts such as

degrees of freedom and model distances (or goodness of fit) are described at

an intuitive level. We also describe and compare the asymptotic distributions

of freedom parameter estimators under product-multinomial and product-

Poisson sampling assumptions. Chapter 3 has more of an applied flavor.

We consider simultaneously modeling the joint and marginal distributions

of multivariate categorical response vectors. A broad class of simultaneous

models is introduced. The models can be fitted using the techniques of

Chapter 2. Several numerical examples are considered. Chapter 4 outlines the

ML fitting technique known as the EM algorithm. This algorithm is used to

fit models with incomplete data. Some advantages and disadvantages of using






-15-

the EM algorithm are addressed. The most important disadvantage is that

the algorithm does not provide, as a by-product, a precision estimate of the

ML estimators. We derive an explicit formula for the observed information

matrix for the Poisson loglinear model parameters when only disjoint sums of

the complete data are observable. An application to latent class modeling is

considered. We also propose an ML fitting algorithm that uses both EM and

Newton-Raphson steps. The modified algorithm should prove to have many

positive features.

In this dissertation, we do not distinguish typographically between

scalars, vectors, and matrices. Parameters and variables are treated as ob-

jects, their dimensions either being explicitly stated or implied contextually.

By convention, functions that map scalars into scalars, when applied to

vectors, will be defined componentwise. For example, if i represents an n x 1

vector, then

logy = (log,lg/A2,...,log~n)'.

We frequently use abbreviations that are common in the statistical

literature. They include ML (Maximum Likelihood), WLS (Weighted

Least Squares), IWLS (Iterative (Re)Weighted Least Squares), and EM

(Expectation-Maximization).

The range (or column) space of an n x p matrix X is denoted by M(X)

and is defined as {p : tz = XP3, 3 E RP}. The symbols and $ are the

binary operators 'direct product' and 'direct sum'. The direct (or Kronecker)

product is taken to be the right-hand product. That is,


A B = {Abij}.






16-
The direct sum, C, of two matrices A and B is defined as


C=A B= A 0).
OB

The symbol D(p) represents a diagonal matrix with the elements of p on the

diagonal. That is,
(1 0 ... 0
D(M) = 0
0 0 ... /n
In Chapter 4, we make use of the bracket notation often used by

statistical and mathematical programming languages (e.g. Splus, Matlab).

To illustrate the notation, consider a matrix A. The (sub)matrix A[, -2] is

then matrix A with the second column deleted. Similarly, the matrix A[-3,]

is the matrix A with the third row deleted.

Equation numbering is consecutive within sections of a chapter, the

first number representing the chapter in which it appears. For example, the

thirteenth equation in section 2.3 is equation (2.3.13). Within each appendix,

the equations are numbered consecutively. For example, the third equation

in Appendix B is numbered (B.3). Tables are numbered consecutively within

chapters so that, for instance, Table 3.2 represents the second table within

Chapter 3. Theorems, lemmas, and corollaries are numbered independently

of each other. All are numbered consecutively within sections. Therefore,

Corollary 3.2.2 is the second corollary within section 3.2 and Theorem 2.3.1

is the first theorem within section 2.3.













CHAPTER 2
RESTRICTED MAXIMUM LIKELIHOOD FOR A GENERAL
CLASS OF MODELS FOR POLYTOMOUS RESPONSE DATA


2.1 Introduction


In this chapter, we consider using maximum likelihood methods to fit a

general class of parametric models for univariate or multivariate polytomous

response data. The models will be specified in terms of freedom equations

and/or constraint equations. These two ways of specifying models will be

discussed at length in section 2.2. The model specification equations may be

linear or nonlinear in the model parameters. Specifically, if p represents the

s x 1 vector of expected cell means, the linear constraints will be of the form

Lp = d and the nonlinear constraints will be of the form U'Clog(Ap) =

0. The freedom equations will have form Clog(Ap) = XPf, where the

components of the vector 3 are referred to as the freedom parameters. In

Chapter 3 of this dissertation, we discuss more specifically models that can

be specified in terms of these constraint and freedom equations. The models

of that chapter allow one to simultaneously model the joint and marginal

distributions of multivariate polytomous response vectors.

The maximum likelihood, model fitting algorithm of this chapter utilizes

Lagrange multipliers and a modified Newton-Raphson iterative scheme. In

particular, the models will be specified in terms of constraint equations and

the log likelihood will be maximized subject to the constraint equations being

17-






18-
satisfied. One common optimization algorithm found in the mathematics

literature is Lagrange's method of undetermined multipliers. We show that

Lagrange's method is easily implemented for ML fitting of the models under

consideration in this chapter. One problem with Lagrange's method of

undetermined multipliers for ML fitting of statistical models has been that it

becomes computationally infeasible for large data sets. By using a modified

Newton-Raphson method which involves inverting a matrix of a simpler form

than the more complicated Hessian, we consider fitting models to relatively

large data sets.

We also explore the asymptotic behavior of the estimators within the

framework of constraint-rather than freedom-models. Usually, asymptotic

properties of model and freedom parameter estimators are studied within the

framework of freedom models. Aitchison and Silvey (1958, 1960) and Silvey

(1959) studied the asymptotic behavior of the model parameter estimators

when the model is specified in terms of constraint equations. Following the

arguments of Aitchison and Silvey, we derive the asymptotic distributions of

both the model and freedom parameter estimators.

Previous work by Haber (1985a) addressed maximum likelihood methods

for fitting models of the form


Clog(A/) = X,,

to categorical response data. Subsequently, Haber and Brown (1986)

discussed ML fitting for loglinear models that were also subject to the

linear constraints Lp. = d, where these constraints necessarily include the

identifiability constraint required of p, the vector of product-multinomial






19-

cell means. Both of these papers advocated the use of Lagrange's method

of undetermined multipliers to find the maximum likelihood estimates of

the model parameters Mp. The method of Haber (1985a) involved using

the (unmodified) Newton-Raphson method which becomes computationally

unattractive as the number of components in p gets moderately large. Both

Haber (1985a) and Haber and Brown (1986) were primarily concerned with

measuring model goodness of fit and therefore did not consider estimation

of freedom parameters. Haber (1985b) did consider estimation of freedom

parameters, but only when the simpler model Clog p = XP3 was used. One of

the several ways that we extend the work of Haber (1985a, 1985b) and Haber

and Brown (1986) is to consider estimation of the freedom parameters when

the more general model Clog Ap = X,3 is used.

Others have considered ML fitting of nonstandard models for multivari-

ate polytomous response data. Laird (1991) outlines the different approaches

taken by different authors. As an example, Dale (1986) considered ML fitting

for a particular class of models for bivariate polytomous ordered response data

which were of the form


C1 log(Al p) = X1fi, g(A2li) = X2,2


Specifically, the first freedom equation specifies a loglinear model for the

association between the two responses measured by the global cross-ratios

(cross-product ratios of quadrant probabilities) so that C1 and A1 are of

a particular form. The second set of freedom equations specifies some

generalized linear model (McCullagh and Nelder, 1989) for the marginal

means or probabilities. Maximum likelihood estimators for the association






20-

model freedom parameters /3 and the marginal model freedom parameters

/32 were simultaneously computed by iteratively solving the score equations

via a quasi-Newton approach. To use this maximization technique, the score

functions, which involve the cell probabilities, must be written explicitly

as a function of the freedom parameter 3 = vec(/l3, 32). A nontrivial

approach to finding reasonable starting values for / is discussed by Dale

(1986). Along with Dale, McCullagh and Nelder (section 6.5, 1989) and

Becker and Balagtas (1991) consider writing the score as an explicit function

of the freedom parameters so that the marginal and association freedom

parameter estimates may be computed simultaneously. In general, when there

are more than two responses, this is not a simple task and so an extension

of this method to multivariate polytomous response data models will be very

messy indeed. Also, convergence of the iterative scheme requires good initial

estimates of the freedom parameter P. These may be very difficult to find. In

contrast, the maximization approach of this chapter, which is similar to Haber

(1985a) and Haber and Brown (1986), is shown to be easily implemented for

fitting multivariate polytomous response data models. With this technique,

it is not necessary to write the cell means as an explicit function of the

freedom parameters. Further, initial estimates of the freedom parameters,

which are difficult to find, are not needed for this technique. Instead, only

initial estimates of the cell means and undetermined multipliers are needed.

Reasonable initial estimates of the cell means are the cell counts themselves.

While a reasonable initial estimate of the vector of undetermined multipliers

is the zero vector-the value of the undetermined multipliers when the model

fits the data perfectly.






21-

We will now introduce the class of models that we will consider for the

remainder of this chapter and the next, more applied chapter. The models

have form


C1 log(A1/) = X18I, C2 log(A2z) = X212, LIp = d


where the linear constraints include the identifiability constraints. Later,

when we study the asymptotic behavior of the ML estimators, we will

require the components of d to be zero unless they correspond to an

identifiability constraint. These models, which are of the form Clog(A/) =

Xfp, Lpr = d, will allow us to model both the joint and marginal distributions

simultaneously when dealing with multivariate response data. The bivariate

association model of Dale (1986) is a special case of these models, as we

can specify the matrices C1 and A1 so that C1 log(Ali) is the vector of log

bivariate global cross-ratios. Restricting the marginal models to have form

C2 log(A2/t) = X2f62, rather than allowing the marginal means to follow a

generalized linear model, as Dale (1986) did, is not overly restrictive. In

fact, many of the generalized linear models for multinomial cell means can be

written in this form. For example, loglinear, multiple logit, and cumulative

logit models are of this form. Also, unlike Haber (1985a) and Haber and

Brown (1986), we will be concerned with estimation of the freedom parameter

3 = vec(f31, 32), thereby allowing for model-based inference.

Model-based inferences usually refer to inferences based on freedom

parameters. With freedom equations, we have the luxury of choosing a

parameterization that results in the freedom parameters having meaningful

interpretations. For instance, a freedom parameter / may be chosen to






22 -

represent a departure from independence in the form of a log odds ratio.

More generally, we usually will try to parameterize in such a way so that

certain parameters will measure the magnitude of an effect of interest.

For example, consider an opinion poll where a group a subjects were

asked on two different occasions whether they would vote for the President

again in the next election. Suppose they were asked immediately after the

President took office and again after the President had served for two years.

The researcher may be interested in determining whether the distribution of

response changed from Time 1 to Time 2 and if so, assess the magnitude of

the change. The data configuration can be displayed as in Table 2.1.



Table 2.1. Opinion Poll Data Configuration
Data Probabilities
Time 2 Time 2
yes no yes no
Time 1 yes yn Y12 Time 1 yes 7rl 7rl12 71I+
no Y21 Y22 no r721 22 72+
7+1 7r+2


We could formulate a model of the form C log(Ap) = X/3 in such a way

so that the freedom parameter 3 has a nice interpretation with respect to the

hypothesis of interest. One such model is

l 2(i)
log ( )=a+pi, i=1,2 (2.1.1)


where the parameter ij(i) is a marginal probability, i.e.

r(i)={ r+, ifi=1
r+j, if i=2






23-

and, for identifiability of the freedom parameters,


P1 = -P2 = P-

Model (2.1.1) is a simple logit model for the marginal probabilities {-rj+} and

{Tr+j}. The parameter p measures the magnitude of departure from marginal

homogeneity in that p = 0 if and only if there is marginal homogeneity.

One could use the Wald statistic P/se(p) to test the hypothesis. If the

null hypothesis is rejected, we can assess the magnitude of departure from

marginal homogeneity by computing a confidence interval for 2p which is the

log odds ratio comparing the odds that a randomly chosen subject responds

'yes' at Time 2 to the odds that a randomly chosen subject responds 'yes' at

Time 1.

This simple example illustrates the utility of using freedom parameters

and the corresponding model-based inferences. For this reason, this chapter

will be concerned with making inferences about both the model parameters

Ap and the freedom parameters 3.

The contents of the following sections are as follows. In section 2.2,

we provide an overview of parametric modeling. The two ways of specifying

models-via constraint equations and via freedom equations-are discussed

at length in section 2.2.1. It is shown that a model specified in terms of

freedom equations can be respecified in terms of constraint equations. In

particular, the freedom equation Clog(Ap) = XP3, which actually constrains

the function C log(Ap) to lie in some manifold spanned by the columns of X,

is equivalent to the constraint equation U'Clog(Ap) = 0, where the columns

of U form a basis for the null space of X'. Other topics covered in section 2.2






24-

include interpretation and calculation of 'degrees of freedom' and measuring

model goodness of fit.

We describe a general class of models for univariate or multivariate

polytomous response data in section 2.3.1. The data vector y is initially

assumed to be a realization of a product-multinomial random vector. We

describe the asymptotic behavior of the product-multinomial ML estimators

in section 2.3.3. Lagrange's method of undetermined multipliers is used to

find restricted maximum likelihood estimates of the model parameters and

the freedom parameters. The actual algorithm is described in detail in section

2.3.4.

In section 2.4, we explore the relationship between the product-multinomial

and product-Poisson ML estimators. General results that allow one to

ascertain when inferences based on product-Poisson estimates are the same as

inferences based on product-multinomial estimates are shown to follow quite

directly when one works within the framework of constraint models. Theorem

2.4.2 of this section, represents a generalization of the results of Birch (1963)

and Palmgren (1981).


2.2 Parametric Modeling-An Overview

Inferences about the distribution of some n x 1 random vector Y are

often based solely on a particular realization y of Y. In parametric modeling

it is often the case that the distribution of Y is known up to an s x 1 vector

of model parameters 8; i.e. it is 'known' that


Y ~ F(y; ), 0e8, E


(2.2.1)






25 -
where 0 is some (s q)-dimensional (q 0) subset of R* known to contain the

true unknown parameter 9*. The cumulative distribution function F maps

points in R" into the unit interval [0, 1] and is assumed to be known.

In general, we will allow the dimension s of 0 to grow with n. For

example, let Y = (Y1,..., Y,) have independent components such that

Yi ~ ind G(yi; zi(O)), i= 1,...,n,

where zi(8) is some function of 0 associated with the ith component of Y.

The function zi could be defined as zi(0) = Oi, in which case s = n. Or, on

the other hand, zi could be a mapping from R' to RI with s fixed.

2.2.1 Model Specification.

In parametric settings, models for the data, or more precisely, models for

the distribution of Y, can be completely specified by recording the family of

candidate distributions that F may belong to. That is, one must specify the

form for F(.; 0) and the space 0M that is assumed to contain the true value

9* of 9. In parametric modeling, the form of F(.; 9) is assumed known, but

the true value 0* is not. Denote a parametric model by [F(.; 9); 0 e OM] or

more simply by [0M]. We say the model [0M] 'holds', if the true parameter

value 0* is a member of 0M, i.e.

[OM] holds 0* e OM.

A model does not hold if 0* V Om.

The objective of model fitting is to find a simple, parsimonious model

that holds (or nearly holds). By parsimonious, we mean that the vector 0 can

be obtained as a function of relatively few unknown parameters. An example






26 -

of a parsimonious model for the distribution of an n-variate normal vector

with unknown mean vector p and known covariance is [Op], where


Op = {A E R" : P/ = a, J = 1,...,n, p unknown}.

Notice that all n components of p can be obtained as a function of

one unknown parameter 3. Thus, all of our estimation efforts can be

directed towards the estimation of the common mean 3. An example of a

nonparsimonious model is the so-called saturated model [O], where


0 = {pI: p E R"} = R".

In this case, p is a function of n unknown parameters.

The question of whether or not the parsimonious model holds is an

entirely different matter. Practically speaking, a model will rarely strictly

hold. Therefore, we will often say a model holds if it nearly holds, i.e. for

some small e

inf |9* 01 < e.

Without delving too much into the philosophy of model fitting and the

simplicity principle (Foster and Martin, 1966), we point out that for a model

to be practically useful it must be robust to the 'white noise' of the process

generating Y. That is, it should account for only the obvious systematic

variation. A model would be said to be robust to the white noise variability,

if the model parameter estimates based on different realizations of Y are very

similar. As an example, if instead of [0E], the saturated model [E] was used

to draw inferences about the normal mean vector pt, we would find that the

model fit perfectly, but that upon repeated sampling the model estimates






-27-
would change dramatically. Thus, the model is not robust to the white noise

of the process. On the other hand, the parsimonious model [Op] estimates

would change very little from sample to sample, varying with the sample

mean of n observations. This model is robust to the white noise variability.

Therefore, if the model would hold, or nearly hold, we would say it was a

good model.

Freedom Models. In the previous n-variate normal example we specified a

model [Op] in terms of some unknown parameter /. Aitchison and Silvey

(1958, 1960) and Silvey (1959) refer to the parameter / as a 'freedom

parameter' and the model [Op] as a 'freedom model'. These labels are

reasonable since we can measure the amount of freedom we have for estimating

9 by noting the number of independent freedom parameters there are in the

model. The model [O(] has one degree of freedom for estimating the mean

vector ~. Thus, once an estimate of the single parameter / is obtained the

entire vector p can be estimated; it is a function of the one parameter 3.

Notice that 'degrees' of freedom correspond to integer dimension in that a

degree of freedom is gained (lost) if we introduce (omit) one independent

freedom parameter thereby increasing (decreasing) the dimensionality of OE

by one.

In general we will denote a freedom model by [Ox], where


ox = {9 e : g(e) = X3,3 E R'}

The function g is some differentiable vector valued function mapping 0 e 0

into r-dimensional Euclidean space Rr. The 'model' matrix X is an r x p full

column rank matrix of known numbers. To calculate degrees of freedom for





28-
[Ox] we will initially assume g satisfies

V00 Ox, (~ ) is of full row rank r.

It also will be assumed that the constraints implied by g(0) = XP3 are
independent of the q constraints implied by the model [O] of 2.2.1. Well
defined models will satisfy these conditions. For example, any g that is

invertible satisfies the derivative condition. Actually this derivative condition
is not a necessary condition for the model to be well defined. Later, we will

show that g need only satisfy a milder derivative condition.

The degrees of freedom for the model [Ox] can be obtained by subtract-
ing the number of constraints implied by [Ox] from the total number of model
parameters, s. The number of constraints implied by [Ox] is (r p) + q, the
dimension of the null space of X' plus the q constraints implied by model [O].
Hence, the model degrees of freedom for [Ox] is

df[Ox] = s (r p + q) (2.2.2)

In view of (2.2.2) the model degrees of freedom, an integer measure of freedom

one has for estimating 9, is an increasing function of p the number of freedom

parameters. In fact, for the special case when q = 0 and g(8) = 0 (so s = r),
we have that the number of degrees of freedom for model [Ox] is simply p,
the number of freedom parameters. This gives us another good reason for

calling f a freedom parameter and [ex] a freedom model.
Constraint Models. Notice that

{0 e E : g(8) = Xp, 3 e RP} (2.2.3)

can be rewritten as


{( e 6 : U'g(8) = 0},





29-

where U is an r x (r p) full column rank matrix satisfying U'X = 0, i.e. the

columns of U form a minimal spanning set, or basis, for the null space of X'.

Letting u = r p and h.(0) = 0 be the q constraints implied by [0], we can

write the (u + q) x 1 vector of constraining functions as h(0) = [hl(0), h,(0)]'

where hi = U'g. We rewrite the freedom model [Ox] of (2.2.3) as [Oh], where

Oh = {0 E R' : h() = 0}. (2.2.4)

Aitchison and Silvey (1958,1960) refer to model [Oh] as a constraint model.

Every freedom model can be written as a constraint model.

We present a few simple examples to illustrate the equivalence between

the two model formulations-freedom and constraint.

Example 1. Let Yi ~ ind N(p3,r2), i = 1,...,n, where r2 is known.

This model can be specified as the freedom model [Ox], where

Ox = {pI E R" : p = lnf, # unknown }

or equivalently it can be expressed as the constraint model [Oh], where

Oh = {/ E R" : U'p = 0}

and U' is the (n 1) x n matrix
1 -1 0 0 ... 0
U'= 0-1 0 -1 0 0

1 0 0 0 .- -. 1
It is easily seen that Ox = Oh and that the model degrees of freedom is

df[Ox]= n-(n- 1)= 1.
Example 2. Let Y. ~ ind N(i( = fio +/3ix,,U2), i = 1,...,n, where a2

is known. This model can be specified as the freedom model [Ox], where


Ox = {A E R" : Pi = 3o +31Xi, i = 1,..., n }






30 -
or assuming that each xi is distinct, as the constraint model [Oh], where


Oh = { E R" : U'l = 0 }.


Here U' is the (n 2) x n matrix

-1 1 + 1 -1 0 ..- 0 0 0
22-Z-1 2-21 Z3-22 Z3-Z2
-1 1 1 -1 0 0 0
22-21 Z2-ZX Z4-23 24-23
U' =

,-1 1 0 0 ...0 1 -1
21-Xl x2-z- Z--1 ZRn-n-1~ /

Notice that U'p = 0 implies that

Aj+I Pi Pk+1 ik
vk, j.
xj+1 xj Xk+1 Xk

That is, the n means fall on a line. As before, it can be seen that Ox = Oh

and that the model degrees of freedom is df[Oh] = n (n 2) = 2.

Definitions. We will assume that the constraining function h satisfies

some reasonable conditions so that the model is well defined. We first present

some definitions.

(1) A model [Oh] is said to be 'consistent' if Oh 0.

(2) A consistent model [Oh] is said to be 'well-defined' if the Jacobian

matrix for h is of full row rank v = u + q at every point in Oh. That is,

0 Oh, h(0) \ of
VOo E h, ( cOh is of full row rank v.


(3) A model [Oh] is said to be 'ill-defined' if it is not well-defined, i.e.


3o E hA, (O) is not of full row rank v.
( ow -I1 0






31 -
(4) An ill-defined model [Oh] is said to be 'inconsistent' or 'incompatible'

if Oh = 0.

Briefly, any reasonable model will have a nonempty parameter space and

hence will be consistent. The Jacobian condition of definition (2) is similar

to the condition required in the Implicit Function Theorem (see Bartle, 1976).

Basically, this condition requires the constraints to be nonredundant so that,

at least theoretically, the constraint equations can be written uniquely as

a function of a smaller set of parameters. An ill-defined model has been

specified with a redundant set of constraint equations. Using the lingo of

the optimization literature, two constraints are redundant if, for each point

in the parameter space, both of the constraints are 'active' or both of the

constraints are 'inactive'. That is, for all parameter values, if one constraint

is active (inactive) then the other is necessarily active (inactive).

It should be noted that the above definitions are in terms of the

constraint formulation of a model. This is sufficient since freedom models can

be written as constraint models. For convenience, we give sufficient conditions

for a freedom model to be well-defined.

A consistent freedom model is well-defined if it satisfies the following two

conditions:

(i) The constraints implied by g(e) = XP3 are independent of the q

constraints implied by [O].

(ii) The Jacobian matrix of g evaluated at any point in [Ex] is of full row

rank r, i.e.


Vo E Ox, ( g(0) is of full row rank r.
(90






32 -

The sufficiency of conditions (i) and (ii) can be seen by observing that

(ii) implies that hi = U'g has a full row rank Jacobian since U' is of full row

rank and (i) implies that h = (hi, h.)' has full row rank Jacobian. These

sufficient conditions are by no means necessary for a model to be well defined

as the Jacobian of h may be of full row rank v even when the Jacobian of g

is not of full row rank.

Notice that the model matrix has nothing to do with whether or not a

model is well defined. In particular, one may think that the model [Ox] is

ill-defined whenever the r x p matrix X is not of full column rank; i.e. the

freedom parameters are nonestimable. However, the model can be rewritten

as a constraint model with the full column rank matrix U spanning the null

space of X, which has dimension less than p r. It follows that if g satisfies

(i) and (ii), then the model [Ox] will be well-defined. The only reason we

have taken X to be of full column rank is to avoid using generalized inverses

when working with the freedom parameters.

To illustrate the use of these definitions, we consider the model [OM],

where


OM = { e Rn : MO d = 0}.



The model will be well defined if Oh/80' = M is of full row rank. It is

inconsistent if the linear system of equations MO = d is inconsistent.

If a model [Oh] is well defined, then the constraints implied by the model

are all independent in that no constraint can be implied by the others. We

will consider only well-defined models when calculating degrees of freedom.






33 -

As before, we calculate degrees of freedom for a model as the difference

between the number of model parameters s and the number of independent

constraints v implied by the model, i.e.


df[Oh] = s (r p + q) = s (u + q) = s- v


Notice that for the constraint model, model degrees of freedom is a decreasing

function of the number of independent constraints v.

Finally, it should be noted that models may be specified in terms of

both freedom equations and constraint equations. In fact, in subsequent

sections this will be the case. However, without loss of generality, we will

concentrate on constraint models since any model can be written in the form

of a constraint model.


2.2.2 Measuring Model Goodness of Fit

Inferences about model parameters are reliable only if the model is

'good'. A good model should be well defined (or at least consistent). It

should be simple and parsimonious. Finally, the model should be relatively

close to holding.

To assess whether or not the model holds, we will need the concept of a

distance between two models. To begin, we will assume there is some measure

of distance between two hierarchical parametric models. (Two models [O1]

and [02] are hierarchical if 02 c 01 and df[O2] < df[O1] whenever 01 02.)

This parametricc) distance will be a quantitative comparison of how close

the two models are to holding. Thus, if both models hold the distance is

zero. The distance will also be independent of the model degrees of freedom.





34 -
Recall that the form of F(.; 0) is assumed known. Therefore, the distance will

measure how far the true parameter is from falling in the parametric model
space. Suppose, firstly, that 01 and 02 are general parameter spaces. That
is, 0 E 01 u 02 does not necessarily define a probability distribution. In other
words, 9 need not fall in a subset of an (s- 1)-dimensional simplex. Let a(8)
and b(8) be vector or matrix valued functions of the unknown parameter 8.

Define a distance between two hierarchical models [01] and [02] (02 C 01) as


6[02; 01] = inf lib(9)(a(9) a(0*))1|2 inf I|b(9)(a(O) a(O*))112.
02 01

Notice that a and b can be chosen so that

(1) 6[02; 01] 0
(2) [02; 01] = 0, iff O1 and 02 hold.
For example, consider the case Y ~ MVN,(i, a2I,). Suppose that

[] = {((p,2) :L E Rn,a2 > 0}

[01] = {((, 2) : l= o,2, > 0}

[02] = {(t, ,2) : X = Xp, p RP, ,2 > 0}

[03] = {(I, 02) : = 1,a, a e R, a2 > 0}.

In this example, each component of Y has a common variance a2. It seems

reasonable that differences between any pj and the true mean t* are equally
important. Hence, a natural distance between any two of these models is


S[0M,; 2M; = inf II. _,*2 inf IIt L *2.
Sn a

Notice that a(i,, ,2) = l and b(f, a2) = 1. Hence, the measure of distance





- 35 -


between [0] and [01] is


6[01; 0] = inf llp *11' = I|0o *ll12

The second infimum is zero since the model [0] is known to hold.

The measure of distance between [02] and [0] is

b[02; ] = inf Ip p *112 = inf IIXP 112

= IIX(XIX)-IXI',* 1*112
(2.2.5)
= I(I. X(X'X)-lX')1*112

= P*'(I- X(X'X)-IX')1X *.

This is the squared length of the vector orthogonal to the projection of p*
onto the range space of X. Notice that if p* = Xf*, that is 02 holds, then

6[02; 0] = 0.
Finally, the distance between [03] and [02] is

6[03; 02] = inf jIp p*112 inf 11i /1*112
03 2
= *(I- ll)* /1*'(I -X(X'XX)-X') (2.2.6)
n
= /*'(X(X'X)-'X 1n1)

As another example, consider a random vector Y = (Y1,...,Y,)', with
independent components following an exponential dispersion distribution

(Jorgenson, 1989). That is,

Yi ~ indep ED(Ai,oa'), i= 1,...,n,

where the density of Yi, with respect to some measure, has form


fy(y; -, 02) = a(y, 2) exp{ T K(71)} (2.2.7)





36 -

where pi = J'(7y) and var(Yi) = a2r"(1y,). Let V(Iu) = e- "(7y) and

0 = (PI,...,n,oa2)'. Since the components of Y have different variances,
a natural measure of distance is


[SOM,; OM] = inf IIV(L)-1/2(-p *)112 inf I V(p/)-L/2( _*)112. (2.2.8)
OM2 eM1


That is a(0) = p and b(O) = V(p)-1/2. Premultiplying the vector (p p*) by

V(p)-1/2 has the effect of downplaying those differences (p/ i *) when the

corresponding variance is large.

To assess the goodness of fit of a model, relative to another, we can

estimate the distance 6 via some statistic based on the observed data. It

is interesting to note that when 6 = 0, i.e. both models hold, our data-

based estimate of this null distance will be some nonnegative (positive, if

the model is unsaturated) number, reflecting the amount of white noise or

random variability there is in Y. This is so because, if both models hold,

then the only reason that our estimate of distance would be nonzero would

be because Y has some random component. That is, the variability in Y that

is not explained by the model causes the data to fit the model imperfectly.

Let D be an estimate of 6. That is, D[E2; 01] is a stochastic, data-based

estimate of how far apart models [01] and [02] are. Potential candidates

for D are the weighted least squares, likelihood ratio, Wald, deviance, and

Lagrange multiplier statistics.

For example, consider the n-variate normal case and the four candidate

models [0], [01], [02], and [03]. We will assume that both [0] and [02]

hold. In view of (2.2.5) a reasonable estimate of 6[02; 0] can be obtained by






37-
replacing /* by Y, the estimate of p* under model [O], i.e.

n
D[o2; ] = Y'(I X(X'X)X')Y = (Y Y)2.
1
Recall, that since [0E2; ] is known to be zero, D[02; ] serves as our

'estimate of error'.

Similarly, a reasonable estimate of 6[03; 02] can be obtained by replacing

p* in (2.2.6) by Y, the least restrictive estimate of p*, i.e.

D[03; 02] = Y'(X(X'X)-'X' Y = (8 )2

Now 03 C 02 and

df[03] = n +1 (n 1)= 2

df[2] = n+ 1 (n- p) = p + 1.
The degrees of freedom associated with estimating the distance between

two models will be called the distance (or residual or goodness-of-fit) degrees

of freedom. The distance degrees of freedom for the two models [OM1] and

[OM ,] is defined to be the difference between the two model degrees of freedom,
i.e.

df (S[OM; M1]) = df [OM1 df[0M1].

The number of distance degrees of freedom measures the dimensional distance

between the two models, i.e. the difference in dimensions. It measures the

difference in the amount of freedom one has for estimating 0 for the two

models. It seems intuitive that if the degrees of freedom is large, that is the

dimensional difference between the two models great, the significance of the

distance statistic may be difficult to ascertain. This follows since we expect

the fit to be quite different for the two very different models, even when both






38 -

models hold. This is a reflection of both white noise and possibly lack of fit.

Therefore, the distance statistic will tend to be large, even when both models

hold. But for many statistics, a large mean implies a large variance, thereby

making significant findings more difficult. It is for this reason that we say

it is better to concentrate our efforts on relatively few degrees of freedom

to detect lack of fit. That is, one should use the smallest alternative space

possible when testing a null hypothesis.

A more technical argument holds when the test statistic (distance

statistic) is a Chi-square or an F. Das Gupta and Perlman (1974) showed

that for a fixed noncentrality parameter, i.e. fixed distance between models,

the power of the F-test or the Chi-square test increases as the distance degrees

of freedom decreases.

Example 1: Continuing with the n-variate normal example, we see that

df(6[O3; 021) = df[Oz] df[03] = (p + 1) 2 = p 1.

Thus, 03 is of p 1 less dimensions than 02. Now, if we knew ,2 the white

noise variance, we could test Ho : 9* e 03, vs. H1 : 8* 02 03, using the

statistic
D[03; 02] _SS(Reg)
2 2 ,(2.2.9)

which has a X2(p-1) null distribution. However, r2 is not generally known and

we must estimate it. One way of estimating ,2 is by estimating the distance

between [0] and [02], two models that are known to hold, and dividing by

the distance degrees of freedom. Since the distance degrees of freedom is

df [] df [2] = n +1 (p +1) = n -p, we have that the estimate of the white
noise variance is D[02; 0]/(n- p) = SS(Error)/(n p).






39 -

Notice that in the above example the estimate of the parameter 02

was simply the estimated distance between two models that were known to

hold divided by their dimensional distance. Quite generally, when the data

have an exponential dispersion distribution (2.2.7) with common dispersion

parameter r2, the estimated distance between two models that are known to

hold, divided by their dimensional distance gives us an estimate of a2. This is

true when the estimated distance is taken to be the LR, Wald, Deviance, LM,

or the weighted least squares statistics. These statistics are natural estimators

of the weighted distance given in (2.2.8) for the exponential dispersion models.

Now, let us assume that 01 and 02 are each subsets of an (s -

1)-dimensional simplex. For example, with count data, conditional on the

total n, the distribution is often multinomial with index n and parameter

(alternatively, probability distribution vector) 0*. Read and Cressie (1988)

extensively study a family of distance measures called the power-divergence

family. The power divergences have form


(0* 0) (+ 1) O [( 1 ; -o
where IJ and I-1 are defined to be the continuous limiting value as A 0 and

A -1. It is assumed that 0* and 0 fall on an (s 1)-dimensional simplex.

As usual, let 0* represent the true unknown parameter. We define the family

of distance measures between [01] and [02] (02 c 01) to be proportional to


6[02; 01] = 2n{ inf I0'(*,0) inf A(0*, 0)}.
02 01

By properties of IX(O*, ) (Read and Cressie, 1988, pp. 110-113), it follows

that S > 0, with equality if and only if both models hold.






40 -
To estimate 6[[2; 01] based on the data, we note that our least restrictive

guess of 0* is Y/n, the vector of sample proportions. Intuitively, a good

estimate of the quantity 6[02; 01] would be
D[02; 0O] = 2n{ inf IA(Y/n, 9) inf IA(Y/n, 0)}
02 01
2 Y 2 Y ^
[(A+1) -n ) A1] (+1)K[(Y ) 1

where 9) and -' are the 'minimum divergence' estimators obtained by

minimizing IX(Y/n,0) with respect to 0 over 01 and 02 respectively. Read

and Cressie (1988) point out that D[02; 01] is equal to the likelihood ratio

statistic when A = 0. Also, if we assume that [01] holds so that the second

infimum is zero, we have that, for A = 1,

D[02; 1]= (Y n ))2

which is asymptotically equivalent to

D[02; 0,] = ( n ))2

where 9(0) is the maximum likelihood estimator of 6* over the space 02. This

is the Pearson chi-square statistic. Other asymptotically equivalent distance

estimates are the Wald statistic and the Lagrangian multiplier statistic. We

now illustrate these results via examples.

Example 2: Suppose that Y = (Yu, Y12, Y21, Y22) is a multinomial vector.

That is,

(Y11, Y2, Y21, Y22)' ~ Mult(n, (7ri, 712,7r21,7r22)'), with i.y = 1.
i j
Thus, the model that is known to contain the true parameter vector 7r* is [0]

where


O = {7r: 7r'14 = 1,7,j E (0, 1), i, J = 1,2}.





41 -

Notice that is really a 3-dimensional subset (simplex) of (0, 1)4 so that

df[9] =4- 1 = 3.

We wish to test the independence hypotheses

SHo : -rlll 22 = 7r2721, VS.
H : 7r11722 = 7"127"21

Writing the model of interest [o] as

O0 = {7 E E : 7r1122 727r21 = 0}

= {r : 7'14 = 1, 71rnr22 712721 = 0},

we can state the independence hypotheses as

Ho : r e00, vs.
H : 7r e o0.

Now, the model degrees of freedom can be found by subtracting the number

of constraints implied by [o0] from the total number of parameters, which

is 4. Hence, df[0o] = 4 2 = 2. Thus, the distance degrees of freedom or

measure of dimensional distance, is df(b[O0; 1]) = 3 2 = 1.

Two distance (goodness-of-fit) statistics commonly used are the Pearson

chi-square X2 (A = 1) and the likelihood ratio statistic G2 (A = 0). The forms

of these two statistics are


D[Oo; O] = X' = (y nri,o)2
i j n7rij,o

and

D[0o; ] = Ga = 2E yj log( Y-i
i j n7r,,o

where iri,o is the ML estimate of 7rij assuming that model [Oo] holds.

Under the null hypothesis, i.e. if independence truly holds, then the

asymptotic distribution of both distance statistics, X2 and G2, is X2(1).






42 -

Example 3: Continuing with example 2, consider the model [EMH] where


EMH = {7 : 7r'14 = 1, 7rl+ +r+1 = 0}.

This model implies that there is marginal homogeneity, i.e. The marginal

distributions for both factors are the same.

We would like to test the hypotheses

Ho : r e O)MH, vs.
H, : 7r EMH.

The model degrees of freedom is df[OMH] = 4 2 = 2, and so the distance

degrees of freedom is df (6[OMH; ]0) = 3 2 = 1. Once again, to illustrate

what model degrees of freedom means, we observe that if [OMH] holds and

we specify two of the four probabilities, the remaining two are completely

determined. Thus, we are free to estimate two of the probabilities based on

the data. The other two are determined.

Two frequently used estimates of the model distance, or model goodness

of fit are the likelihood ratio statistic G2 and the McNemar statistic M2. For

2 x 2 tables, the McNemar statistic and the Lagrange Multiplier statistic are

equivalent since both are score statistics (Agresti, 1990; Aitchison & Silvey,

1958). The statistics take the following forms


D[OMH; O = G2 = 2 = E 2log( Yij
jn ij,o
i jfji,0
and

D[eMH;e] = M- (2-2
Y12 + Y21
where the iij,o in the first expression is the ML estimate of 7rij under the

model [OMHI.






43 -
Under the null, i.e. when the marginal distributions are homogeneous,

both of these statistics have asymptotic X2(1) distributions.

It is important to note that, had the constraint 72+-Tr+2 = 0 been added,

the model would remain consistent but would be ill defined. For 2 x 2 tables,

this additional constraint is exactly the same as the constraint 7r+ r+1 = 0.

2.3 Multivariate Polytomous Response Model Fitting

In this section, we describe ML model fitting for an integer valued

random vector Y that is assumed to be distributed product-multinomially.

We also investigate the asymptotic behavior of the ML estimators within the

framework of constraint models. The models we will consider have form

Ox = {( E O: Clog(Ae)) = XP3, Lee = 0}

or equivalently, for appropriately chosen U,

Ox = Oh e E : U'Clog(AeC) = 0, LeC = 0},

where ee is the s x 1 mean vector of Y, a product-multinomial random vector

and the model parameter space O is of dimension s q, where q is the number

of identifiability constraints. We use the parameter rather than 11 = ee

for several reasons. One reason will become evident when we explore the

asymptotic behavior of the ML estimator of It turns out that the random

variable 4 po is not bounded in probability, whereas 6o is. In fact, the

random variable o converges in probability to 0. Another reason for using

rather than y is that the procedure for deriving the maximum likelihood

estimate of is less sensitive to small (or zero) counts. The range of possible

values is the whole real line, while the range of possible p values is restricted






44 -

to the positive half of the real line. By using 6 the problem of intermediate

out of range values (e.g. negative cell mean estimates) is avoided.

As stated above, we initially assume that the vector of cell counts Y

has a product-multinomial distribution. This is not overly restrictive since it

will be shown that inferences based on maximum (multinomial) likelihood

estimates are often the same as inferences based on maximum (Poisson)

likelihood estimates. We will present some results in section 2.4 that allow

us to determine when these inferences are indeed the same.

We also consider an alternative method for computing the maximum

likelihood estimators and their asymptotic covariances. The method of

Lagrange undetermined multipliers is well suited for maximum likelihood

fitting of the models we will be considering. This is so because we will specify

the models in terms of constraint equations and the fitting problem will be

one of maximizing a function, namely the log likelihood, subject to some

constraints, namely that 6 E Oh.







2.3.1 A General Multinomial Response Model




In this section we specify a class of models that is directly applicable

to Chapter 3 of this dissertation. Specifically, the models will be specified in

such a way so as to include the class of simultaneous models for the joint and

marginal distributions considered in Chapter 3.






45 -
Let the random vector Y = vec(Yi,..., YK) denote a product multinomial

random vector, i.e.


Yi= (Yil,...,YiR)' ~ ind Mult(ni, ri), i = 1,...,K, K > 1,


where the R x 1 vector of cell probabilities satisfy 7rilR = 1, i = 1,..., K.

Consider the 1:1 reparameterization from {Ir;} to {(~}, where &, =

log(pi) = log(ni-ri) is an R x 1 vector of log means. Under this parame-

terization,


Yi ~ ind Mult(ni, ), e41=ni, i= 1,...,K,


or

Yi ~ indMult(ni, ), i=1,...,K, e'(ef1R) = n', (2.3.1)
ni

where n' = (nl,..., nK) is the 1 x K vector of multinomial indices.

The kernel of the log likelihood for Y, written as a function of e, is


e(M)(; y) = y'e, e'(e$ 1R) = n' (2.3.2)


We now posit a model for the vector of log means. Let s = RK be the

total number of cell means. Our objectives are to test the model goodness

of fit and to estimate the s x 1 model parameter vector as well as any

freedom parameters of interest. It will be assumed that the model [ex] can

be specified as

Ox = {( e R' : C1 log Alet = Xi/3, Ca log A2e = X2,2, Lee = 0,
(2.3.3)
e'( 1R) =n',






- 46 -


where
Ci = (fCij, Cij = Cil, is qi x mi i = 1,2

Ai = qfAij, Aij = Ail, is mi x R, i= 1,2

L = 'Lfj, Lj L1 is dx R

= vec(,..., ) and is R xl 1

Xi is Kqi x pi of full rank pi, i = 1, 2

n is the K x 1 vector of multinomial indices

s = RK, the total number of cells
Let us say that a model that can be specified as in (2.3.3) satisfies

assumption (Al). That is,

(Al) The multinomial response model can be specified as in (2.3.3).

Notice that the K matrices of Ci are all identical, likewise with the

matrices comprising Ai and L. This requires that the model does not change

across the K populations (K multinomials). Also, the two sets of freedom

equations in (2.3.3) will allow us to use two different types of models for

the expected cell means. This provides us with enough generality to fit

many interesting models. For example, we may wish to simultaneously fit

a linear-by-linear association loglinear model for the joint distribution and a

cumulative logit model for the marginal distributions.

We can conveniently rewrite (2.3.3) as

Ox = { e R' : Clog(Aee) = XP, LeC = 0, eC'(ef1R) = n', (2.3.4)

where A'= [A', A'], C = C1 ) C2, X = X1 X2, and 3 = vec(3l,3Q2).

Notice that the model [Ox] is specified in terms of both freedom

equations and constraint equations. We will rewrite [Ox] as a constraint






47 -
model keeping in the back of our minds that the freedom parameters may be

of interest also.

Let U be a K(ql + q2) x u matrix of full column rank u such that
U'X = 0. Here u is the dimension of the null space of X', A((X'), i.e.

u = K(qi + q2) (1 +p2). Since U can be chosen to be of full column rank, it

follows that the columns of U form a basis for the null space of X'. Thus, the

range space of U equals the null space of X', i.e. M(U) = A(X'). Multiplying

the right and left hand side of the freedom equation Clog(Aee) = XP/ by U',

we can rewrite (2.3.4) as


Oh = { e R' : U'Clog(Aee) = 0, Lee = 0, e'(elR) n' = 0}. (2.3.5)

Thus, Ox = Oh and the models [Ox] and [Oh] are one and the same.

At this point, we will assume that the constraints implied by the model

[Oh] are nonredundant so that the model is well defined. More specifically, let
h'() = [(U'Clog(Ae))', e'L'] be the 1 x (u + 1) (1 = Kd) vector of constraint

functions. We will assume that the u ++ K constraints implied by h(() = 0

and ee'(@IelR) = n' are nonredundant. Notice that the constraints in h(() = 0

do not include the identifiability constraints. We treat the identifiability

constraints separately for reasons that will become apparent when we actually

fit the models.

As stated previously, one of our primary objectives is to estimate the

model parameters 6 and the freedom parameters f under the assumption
that [Ox] (and [Oh]) holds. We will use the maximum likelihood estimates,

which can be found by maximizing the log likelihood of Y subject to the

constraint that [Oh] holds.






48 -

The (kernel of the) log likelihood under the product multinomial

assumption is shown in (2.3.2). It is


fcm) (; Y) = Y1,*


Thus, we are to maximize the function e(M)(E; y) = y' subject to e OEh.


2.3.2 Maximum Likelihood Estimation

In this section we will discuss two procedurally different approaches

to maximizing the log likelihood e(M)(; y) subject to E e ,. The first

approach, which is the more commonly used approach, requires that the

model be specified entirely in terms of freedom equations. Often times,

when there are no identifiability constraints, the model can be completely

specified as a freedom model. Models amenable to this approach include the

Poisson loglinear model and the Normal linear model. The second approach,

Lagrange's method of undetermined multipliers, can be directly applied when

the model is specified completely in terms of constraint equations. Since the

product multinomial model includes identifiability constraints, it can more

easily be specified in terms of constraint equations. For this reason this

second method is the preferred choice. In the following sections, we discuss

some additional features of these two methods.

Freedom Parameter Approach. One approach often used in simple situa-

tions, namely those situations when the model can be specified completely

in terms of freedom equations, is to write the parameter C as a function

of the freedom parameter I and maximize e(M)((P); y) with respect to 3.

The vector (P3) will be in the model space, since the model was specified





49 -
completely in terms of f/. For example, if the model could be specified as


Ox = {( E R' : log e = Xp},

then ((/) = XP3. Notice that the multinomial model, which includes the K

constraints eV'($flR) = n', is not directly amenable to this approach. In fact,
we would have to reparameterize to a smaller set of s-K model parameters

that account for the K constraints. This reparameterization results in an

asymmetric treatment of the e and for that reason is deemed undesirable.

On the other hand, the Poisson model considered below, will often lend itself

to this maximization approach, since the K constraints eC'(e$1R) = n' are

not included.

Computationally, the method of maximizing the log likelihood with

respect to the freedom parameters is usually simple. Assuming the log

likelihood is concave and differentiable in 3, we need only solve for the root

of the 'score equations', viz.


s(; Y) = ; ) .

Many of the asymptotic properties of the maximum likelihood estimator

3 for 3 are derived by formally expanding the score vector s(P3; y) about the
true value f = 3* in a linear Taylor expansion. That is,


s(/; y) = s(3*; y) + Os(,*;y P ) ) + ( 12) (2.3.6)

In particular, in many situations,


O=s( ; Y)=s(3*;Y)+ Os' () p*) + Op(1),
0/3'






- 50-


so that / 3* has the same asymptotic distribution as

SY) s(3 *; Y).


Subsequently, we will derive the asymptotic distribution of3 -P3* in a different

way. This alternative derivation of the asymptotic distribution of the freedom

parameter estimate will shed new light on the relationship between the

asymptotic behavior of the estimates under the two sampling assumptions-

product Poisson and product multinomial.

Expression (2.3.6) also gives some indication of how one might numer-

ically solve for /, the root of the score equation. A Newton-Raphson type

algorithm is often used. This root finding algorithm involves the inversion

of the derivative matrix as(P3;y)/Ql3', which is usually of small dimension

since the model is usually specified in terms of a small number of freedom

parameters. In fact, the dimension of the derivative matrix will not be larger

than s x s, which occurs when the model is saturated.

Constraint Equations Approach. In many situations, it may be difficult to

specify a model in terms of only freedom parameters or perhaps it is possible

but the researcher would like to treat the model parameters symmetrically,

which would necessitate an additional constraint equation. It also could be

that the function ClogAeE is not a 1:1 function of so that for given /, we

can not solve for explicitly. In any of these cases, we may not be able to

use the aforementioned maximization approach.

In this section, we consider an alternative method for finding that i

that maximizes the function e(M)({; y) subject to E Oh. The method we

will use is the Lagrange's method of undetermined multipliers. Aitchison and






51-
Silvey (1958, 1960) and Silvey (1959) provide much of the essential underlying

theory related to this approach. Three positive features of this method

include (i) estimation of both ( and 3 is possible, (ii) the method provides

us with another enlightening way of deriving the asymptotic distribution

of the freedom parameter estimators, and (iii) the method works quite

generally. A negative feature of this approach is the computational difficulty.

Computationally, the method becomes burdensome as s, the number of log

mean parameters, and u + 1 + K, the number of constraints implied by the

model, become large. In fact, the algorithm involves the inversion of an

(s + u +1) x (s + u +1) matrix. One positive note, is that this potentially very

large matrix does have a simple form and one can invoke some simple matrix

algebra results to reduce the inversion problem to one of inverting matrices

of dimensions (u +1) x (u +1) and s x s.

To best illustrate the difference in computational difficulty of the two

methods, we consider the following normal linear model example. Let


Yi ~ ind N(;i = 3o + alxi,

The log likelihood can easily be written as a function of f = (Po, /i)'.

Maximizing this likelihood with respect to f involves working with a 2 x 2

matrix. On the other hand, we could equivalently specify the linear model in

terms of the 98 constraints,

Pi+1 Pi Pi+2 i+1, i= 12, ... 98,
Bi+i Xi +i+2 Zi+l

and use Lagrange's method. In this case, we would need to invert a matrix

which has dimension (s + u + 1) x (s + u + 1) = 198 x 198.





52 -
Even when we use the matrix algebra results that simplify the problem
of working with the 198 x 198 matrix, we still are left with a formidable task.
It seems that when s is large and the model is parsimonious, i.e. u + + K,
the number of constraints is large, the undetermined multiplier method may
not be the method of choice. However, in time, as computer efficiency gains
are realized, we predict that the scope of candidate models to be fit using
this method will increase tremendously. In fact, at present, many categorical
models can easily be fit using Lagrange's method. We discuss in more detail
how we can use the method of undetermined multipliers to fit models like

[Oh] of (2.3.5).
We are to maximize the function (M) ($; y) = y', subject to the constraint

( E Oh, where
Gh = {~ e R : U'Clog(Ae4) = 0, Le4 = 0, e'($jflR) n' = 0}

= { R: h() = 0,et'(flR)= n'},
and h'({) = [log(e4'A')C'U, eE'L'].

Consider the Lagrangian objective function

F(7) = e(M)(6; y) + (et'(eKlR) n')7 + h'(\)A,

where 7 = vec(, r, A). The K x 1 vector r and the (u + 1) x 1 vector A are
called either 'Lagrange multipliers' or 'undetermined multipliers'.
Provided a maximum exists and that the Jacobian of [e6'(eK1R)-

n', h'(()] is of full row rank u + 1 + K for all 6 e Oh, we can solve for the
maximum by solving the system of equations

F () + D(e'')( lR)(') + H(iM)) )
% = ( f@ 1 )e m) -'n = 0 (2.3.7)
7 /h((M))






53-

where the matrix H() = 8h'(()/98. The Jacobian condition basically

requires the constraints to be nonredundant, thereby making [Oh] a well-

defined model.

From this point on, for notational convenience, the indices for the direct

sum will be omitted unless they are different from 1 and K.

We now require the matrices of models [Ox] and [Oh] to satisfy some

additional conditions. Let us assume that

(A2) Either C = Iq,K or Ci( lm,)= 0, i = 1,2

and

(A3) If C = Iq,K then M(Xi) D M(l$m,)




The assumptions require Ci to be either a contrast matrix (rows sum

to zero), a zero matrix, or the identity matrix. If Ci is the identity matrix,

it will be required that there exists a set of columns in Xi that spans a

space containing the range space of $(Klm,. For most models of interest

these conditions are met. For example, any of the logit type models, such as

cumulative or multiple logit models, can be specified with C being a contrast

matrix. For loglinear models, the condition (A3) is met whenever the model

includes a parameter for each of the K multinomials.

The following lemma will be useful in showing that the maximum

likelihood estimates of and j3 are equivalent under both sampling schemes-

product-Poisson and product-multinomial. The lemma will also enable us to

reduce the number of equations in (2.3.7) that must be simultaneously solved

when computing the maximum (multinomial) likelihood estimators.





54-
LEMMA 2.3.1. If the matrices of models [Ox] and [Oh] satisfy (Al), (A2),
and (A3), then provided the model holds

( -) -= ( 1'~)H() = 0.


Proof. Using matrix derivatives (MacRae, 1974; Magnus and Neudecker,
1988), it follows that

H(s) = [D(ef)A'D-I(Ae4)C'U, D(ee)L']

Thus,

(e 1')H() = [(e e)A'D-1(Aee)C'U, ( e:)L]

= [(@ee)[A',A']D-1 ) (C e( C()U, @eeL.]

[[( e et)A'D-1(Alef), (e eci)A'D-1(A2e')](C. e C2)U, 0]

= [[( e eA'i)D-'(Alef)C' ( eeA'i)D-1(A2e)C2U, o]

= [( e 1')c, ( 1m, )CE]U, O

= 0,0]
=0
The third equality follows since the model holding implies that seiL = 0.
The sixth equality can be seen via the following argument.
If both Ci's are contrast matrices, or zero matrices, then (A2) implies
that the matrix [( $11)C', ($1',)C'] is the zero matrix. On the other hand,
if both C1 and C2 are identity matrices, then since the columns of U span
the null space of X', which, by (A3), implies that the columns of U span a
set contained in the null space of

elm2 '
sim, '





55-
we have that [(e 1m), ( 1'))]U = 0. Any other combination of CO and C2

can also be seen to result in the matrix equaling zero. 0
The following theorem gives conditions under which we can find the ML
estimators of 6 by solving a reduced set of equations. The smaller system of
equations no longer includes the identifiability constraint equations.
THEOREM 2.3.1 Let vec((M), i(M), k(M)) be the solution to (2.3.7).
Assuming that (Al), (A2), and (A3) hold, the sub-vector vec(&(M), \(M))
is the solution to the reduced set of s +u + equations

h(+ H((M))0 (2.3.8)



Proof: Premultiplying the first set of equations in (2.3.7) by $1'W, we arrive
at


( 1'l)y + ( 1'I)D(eZM)( 1R)T + ( l1')H((M))iAM) = 0 (2.3.9)

Now, (e l')y = n and (E l'm)D(eE(M) = e~M)'. Also, since (M) E Oe it must
be that ( eeM )( e 1R) = D(n), the diagonal matrix with the multinomial
indices on the diagonal. Further, by Lemma 2.3.1,
( 11 ')H( /(M)) = 0. Therefore, (2.3.9) can be rewritten as


n + D(n) i(M) = 0,

which implies that 4(M) = -1K. Now, since the identifiability constraints have
been explicitly accounted for when solving for f(M), we can replace i(M) of
(2.3.7) by -1K and omit the identifiability constraints. Thus, vec( (M), \(M))





56 -

is the solution to the reduced set of equations

( (m) + H(W(M))A(M) =

This is what we set out to show.

Before detailing the iterative scheme used for solving (2.3.8), we will

explore the asymptotic behavior of the estimator 0(M) = vec((M), ^(M))

within the framework of constraint models.

2.3.3 Asymptotic Distribution of Product-Multinomial ML Estimators

In what follows, we will assume that K, the number of identifiability

constraints, is some fixed integer, K > 1. We also will assume that the

asymptotics hold as n. = min{ni} approaches infinity and that n. ~ ni, i =

1,..., K. That is, we assume that the asymptotic approximations hold as

each of the multinomial indices get large at the same rate.

The derivation of the asymptotic distribution of b(M) will follow closely

that of Aitchison and Silvey (1958). Briefly, Aitchison and Silvey show that

if the score vector is op(n) and the constraints are such that the derivative

matrices H(() and OH'(()/98 have elements that are bounded functions then,

provided certain mild regularity conditions hold, the maximum likelihood

estimator is an n-1/2-consistent estimator of o and A is an n1/2-consistent

estimator of 0. They show that the joint distribution of (n1/2( -o), n-1/2)

is multivariate normal with zero mean and covariance matrix

(B- B-1H(H'B-H)-H'B-1 0 (
0 (H'B-'H)-1)

where B is the information matrix and H is the derivative of the constraint

function.






-57-

In our application, however, there are some minor changes. With the pa-

rameterization we use, the information matrix is zero since the (multinomial)

log likelihood (2.3.2) is linear in the parameter This happens because the

identifiability constraints eE'( of 1R) = n' are ignored, to preserve symmetry,

when differentiating. Also, in our parameterization, the constraints are in

terms of ee, the components of which are eCi- = n7rij. Thus, the constraints

and the corresponding derivative matrices may not be bounded. For example,

a typical constraint is of the form Let = 0. It follows that the components

of Let and the derivatives are increasing without bound as the multinomial

indices are allowed to increase without bound.

Fortunately, we can still use the results of Aitchison and Silvey (1958)

by replacing the matrix H and the vector A/n of Aitchison and Silvey by

our H/n. and A, where n. = min{ni}. The zero information problem can be

solved by identifying the vector Y e as the 'score vector'. It is pointed out

that, in this case, the asymptotic variance of @D-'/2 (nlR) times the score

vector is not equal to the negative derivative matrix D(-ro) but instead is

equal to D(ro) E-roi7r'j. This happens because the components of Y are not

independent; Y is product multinomial. Using this reparameterization, all of

the necessary assumptions required by Aitchison and Silvey (1958) hold, i.e.

assumptions X and of Aitchison and Silvey (1958) hold.

As previously mentioned, Aitchison and Silvey show that A is an

n1/2-consistent estimator of 0. With our paramterization, having replaced

A/n by A, it follows that A(M) will be n,1/2-consistent. We now derive the

asymptotic distribution of b(M).





58 -
Define the stochastic function g by

g(O; Y) Y eY + H())

The maximum likelihood estimator 0(M) is the solution to g(O; Y) = 0.
Under our parameterization, using the results of Aitchison and Silvey
(1958), we have that each of the following hold

e(M) e6o = D(el)( (M) ) + Op(1),

H( (M)) = H(0) + Op(nl/),


h((M)) = h(o) + H'(Eo)( (M) o) + Op(1)

= H'(-o)( (M o) + Op(1),
and
H(J(M))^(M) = H()A(M) + Op(l).

Thus,

O = g(m); Y) Y em) + H( (M))(M )

can be rewritten as

0 =Y eo D(eo)( (M) _o) + H(O )Op(l1)
H'( o)( (M) o) + Op(1)
(Y e)o (D(eo) -H( o) (M) 0o O (1)
V 0 -H'() 0 M(M)
Therefore, it follows that

eD-1/2(n,1R) (Y eo) =

D(ir) niO~l n /2(+o(n;'/' (2.3.10)

since n, ~ n, i= 1,...,K and 0ro = ( D-'(nil1))ef0.






59-
Now, the random variable SD-1/2(nilR)(Y-e6o) is a vector of normalized
sample proportions so that

( D-1/2(nl1R)(Y- e(o))


has an asymptotic normal distribution with zero mean and covariance matrix

(D(7ro) -ED70,7r1, 0)
0 0'

Therefore, by an extension of a theorem of Cramer (1949) and by equation
(2.3.10), it follows that n/2( (M)- 0,) = n/vec((M) 0o, i(M)) has an

asymptotic normal distribution with mean zero and covariance

D(ro) -(O D(7ro) e7roi7r o D(7ro) -2
(n. (2.3.11)
_o 0 0 0 _(o 0
S* \ *

This covariance matrix is shown in the appendix to have the simple form

(M, 0
0 M)

where


M, = D--'(oo) D-)1()H(H'D-'(o)H)-'H'D-'(7o) $fK 1Rl


and

M2 = n (H'D-( ro)H)-.

Finally, using the fact that n, ~ ni, i = 1,..., K, we can discriminantly
replace n* by the appropriate n, to arrive at a simple, asymptotically
equivalent, expression for the asymptotic covariance of i(M) = vec((M), \(M)).





- 60 -


It is

D-1 D-'H(H'D-'H)-'H'D-1 R 0
0 (H'D-1H)-1) '(

where D = D(po) = D(eo) and H = H(o).

2.3.4 Lagrange's Method-The Algorithm

In this section, we give details of how one can actually fit the models
of (2.3.4) or equivalently (2.3.5). We show how Lagrange's undetermined
multipliers method can be used in conjunction with a modified Newton-
Raphson iterative scheme to compute the ML estimators and their asymptotic
covariances. We will assume that the model assumptions (Al), (A2), and (A3)
hold. This section includes an outline of the algorithm used in the FORTRAN
program 'mle.restraint'.
Recall that our objective is to find that (M) e Ox, where

Ox ={ R': Clog(AeC)=Xf3, Lee=0, (el))e=n},

that maximizes the multinomial log likelihood


(M) (; y) = Y'

Since the assumptions (Al), (A2), and (A3) hold, we see by Theorem
2.3.1 that our problem is reduced to one of solving the system of equations
(2.3.8), i.e. to find the ML estimator 9(M) = vec(i(M), \(M)) we must
simultaneously solve the system of s + u +1 equations


( -e-Y e4H()A =0,
g~o) = ( h()





61-

where the (u + 1) x 1 vector h and the s x (u + 1) matrix H are defined as

follows.

h()= U'Clog(Ae$)

and
Oh' (()
H() =

It will be shown in section (2.4) that g(0) is actually the derivative

of the Lagrangian objective function under the product-Poisson sampling

assumption.

The iterative scheme used in the FORTRAN program 'mle.restraint' is

a modified Newton-Raphson algorithm. The algorithm can be sketched as

follows.


(1) Find a starting value for 8.

(2) Replace 0(") by 0("+1) = O(V) G-1((Y))g(o(")) (2.3.13)

(3) If ||g(0(v+l))|l > tol go to (2). Else stop.


The matrix G(8) used in step (2) is actually


G() + Op(n1 /2) (-De) H( ))


and the inverse of G(O) is of the very simple form (see Aitchison and Silvey,

1958 or Rao, 1974)

G-'() _D-1 D-1H(H'D-1H)-'H'D-1 -D-1H(H'D-1H)-1
-(H'D-'H)-1H'D-1 -(H'D-1H)-
(2.3.14)






62 -

where D = D(ee). Since we use G(0) in place of the Hessian matrix, the

procedure is a modification to the Newton-Raphson method. Haber (1985a)

used the more complicated Hessian matrix.

Notice that the inversion of G, which may be performed at each iteration,

is not nearly as difficult as inverting a general matrix of dimension (s + u +

1) x (s + u + 1). First of all, in view of (2.3.14), to obtain the inverse of the
partitioned matrix G, we need only invert the matrices D and H'D-1H, which

are of dimension s x s and (u + 1) x (u + 1). Secondly, the inversion of D is

simple since D is a diagonal matrix with et on the diagonal. Hence, the most

formidable task in the inversion process is the inversion of the symmetric

positive definite matrix H'D-1H. There are many efficient ways to invert

large symmetric positive definite matrices.

Upon convergence of the algorithm (2.3.13), estimates of the asymptotic

covariances of (M) and A(M) are readily calculable. Write G-1() of (2.3.14)

as



where
P = D-1 D-1H(H'D-1H)-1H'D-1

Q = -D-IH(H'D-1H)-1

R =-(H'D-2H)-1
By (2.3.12), the asymptotic covariance of i(M) = vec((M), (M)) can be

estimated by

var("l)=( P-)efi 0 )

Variance estimates for other continuous functions of ^(M), such as
A(M) = eF(M and t(M) = (X'X)-1X'Clog(Aee~M), can be found by invoking






- 63 -


the delta method. For example,


var(A(M)) aD(eD m ))var(^(M))D(ee) )


and

var( (M)) a

(X'X)-lX' CD- (AA(M))A(var(Af()))A'D- (Ac(M))C'X(X'X)-.

Evidently, Lagrange's method of undetermined multipliers provides us

with a convenient procedure for maximum likelihood fitting of models in a

very general class of parametric models for multivariate polytomous data with

covariates possible. We now briefly outline the steps needed to perform the

iterations of (2.3.13).

Computing U. The first thing we must do is write the freedom model (2.3.4),

which can easily be input by the user, as a constraint model (2.3.5). Therefore,

we must compute a full column rank matrix U that satisfies U'X = 0. The

method we use to find U is attributed to Haber (1985b).

Using the notation of 'mle.restraint', let X be a full column rank matrix

of dimension q x r. Let u = q r be the dimension of the null space of X'.

Further the matrices A and C of (2.3.4) will have dimensions m x s and q x m

respectively. The relationship between these dimension variables and those

used in sections 2.3.1 and 2.3.2 is as follows

q K(q + q2)

r -pi +P2

m K(m, +m2-).


We use the variables q, r, and m for notational convenience.






64 -

Consider the matrix U* = I, X(X'X)-1X'. This q x q matrix is of rank

u = q r and satisfies the property

U*' X = 0.

Let W denote a q x u matrix with random elements. Specifically,


Wij ~ Uniform(0,100), i=l,...,q, j=l1,...,u.

It follows that the matrix W is of full column rank with probability one and

hence that the q x u matrix U = U*W is of full column rank u with probability

one. But the matrix U satisfies

U'X = W'U*'X = W'O = 0.

Therefore, at least with probability one, we have found a full column rank

matrix U that satisfies the property U'X = 0. Using this U, we are able to

write freedom model (2.3.4) as a constraint model (2.3.5).

Computing h(s). We write the constraint model of (2.3.5) as


{( e R : h()) = 0, e6'(e1lR) = n'}, (2.3.15)

where the constraint function h is defined as


h() = (U'Clo(Aet)

Computing g(O). Notice that since (Al), (A2), and (A3) hold, the

identifiability constraints present in the product multinomial model (2.3.4)

can be accounted for explicitly. It will follow by results of section 2.4, that

under either sampling scheme-product-Poisson or product-multinomial-





65 -

the maximum likelihood estimators for ( and A can be found by solving the

equation



g(0)= et H()A) =0, (2.3.16)

where the matrix H is the derivative of h' with respect to (.
Computing H((). We will use matrix derivative results of MacRae (1974)
to find the matrix of derivatives of the constraint function h'().


H(0 h = ()= -[log(e'A')C'U, ee'L']

= [D(ee)A'D-'(Ae4)C'U, D(e4)L'].

The equality follows upon using the matrix version of the chain rule. Notice
that

a aef' a
(log(e 'A')C'U) ( (log(ee'A')C'U)

=D(e) log(eA') )CU + 0
.e ) OAe ]
= D(ee)A'D-1(Ae)C'U

and that

Oet' L' Oee' Oet' L'
S- D(ee)L'.

Computing G(8). The iterative scheme (2.3.13) used to solve the system
of equations (2.3.16) is actually a slight modification of the Newton-Raphson
algorithm. It is a modification because we do not use the derivative matrix
G* = Og(O)/O0 to adjust at each iteration, as Haber (1985a) did, but rather a
simpler matrix G that is related to G* by G* = G+ Op(n2 ). The derivative





66 -
matrix G* can be computed as follows.

G*(8)- = g(O) = [O )

(-D(e) +
\ H'()

H, ()


Og(O) J

H( )


0 0/
+W~) o)
V(" H o o


The matrix
OH( )A __ H()(

is of order Op(n/2) when it is evaluated at 9 = vec(, A) since

H( =
aH Op(n,)
'96


and


A = Op(n.1/2).


It follows that the matrix G, which is much simpler to invert than G*, can
be used to adjust the estimate at each iteration.
Computing the inverse of G. Although the matrix G is of dimension
(s + u + 1) x (s + u + 1), which may be very large in practice, its inverse
is relatively simple to calculate. The inverse of the partitioned matrix

,G (-D H\
= (H'

is shown by Aitchison and Silvey (1958) to have form

G-1 (D-1 D-1H(H'D-H)-H'D- -D-H(H'D- HD- )-1
-(H'D-1H)-IH'D-1 -(H'D-1H)-1 )

Therefore, only the matrices D and (H'D-1H), which are of dimensions
s x s and (u + 1) x (u + 1), need to be inverted. The inverse of D is easily






- 67-


calculated since D is a diagonal matrix with e4 on the diagonal. The inverse

of (H'D-1H), a symmetric positive definite matrix, can be found quite easily,

even when u + 1, the number of constraints, is large. It should be pointed

out that when s, the total number of cell means, is large, the number of

constraints u + 1 may be large and on the same order as s. This will be the

case for parsimonious models-those models with many constraints relative

to number of model parameters.

One could choose to invert the matrix G a limited number of times to

mitigate the computational burden. In fact, in their 1958 and 1960 papers,

Aitchison and Silvey advocate an iterative method whereby the inverse of G

is computed only two times. Once at the initial iteration and again at the

final iteration, upon convergence. We feel, however, that in this special case

in which the matrix G has a particularly simple form, the inverse can be

computed at each iteration. Along with increased computing power, there

are many efficient algorithms for inverting large symmetric positive definite

matrices.

2.4 Comparison of Product-Multinomial and Product-Poisson Estimators

We begin this section by introducing notation for a product-Poisson

random vector.

The sxl random vector Y = vec(Y,,..., YK) is said to be product-Poisson

if

Yij ~ indPoisson(eei), i =1,...,K, j=1,...,R. (2.4.1)

Suppose that the s = RK log means {(ij} satisfy the model [O(p)] where

8) = {( R': Clog(Ae4) = X3, Le = 0}





68 -

or equivalently, for appropriately chosen U,


(P) = P) = { E R' : U'Clog(Aee) = 0, Lee = 0} (2.4.2)


This model implies all the same constraints on ( as the product-

multinomial model [Oh] of (2.3.5), with one exception-the identifiability

constraints, eV'( $ lR) = n', are not included.

Denote the maximum likelihood estimators computed assuming (2.4.1)

and (2.4.2) by (P) and 3(P). Similarly, denote the maximum likelihood

estimators computed assuming (2.3.1) and (2.3.5) by (M) and (M).

Recall that the three product-multinomial model assumptions are

(Al) The multinomial response model can be specified as in (2.3.3).

That is the model parameter space can be represented as

Ox = { E R' : C1 log Ale = XP1, C2 log A2e = X232,

Le = 0, e'(ilR) = n'},

where

Ci = DgCij, Cij Cil, is qi x mi = 1, 2

Ai = afAij, Aij Ail, is mi x R, i = 1,2

L = $ Lj, L =- L1 is dx R

= vec(,,..., Kg), and k is R x 1

Xi is Kqi x pi of full rank pi, i = 1,2

n is the K x 1 vector of multinomial indices

s = RK, the total number of cells.





69 -

(A2) Either Ci Iq,K Or Ci( (l m,)= i =1,2,

and

(A3) If C, = IqK then M(X,) D M(lm,,).

The following theorem states that the maximum likelihood estimators for

E and hence p are the same under the product-multinomial sampling scheme

of (2.3.1) and the product-Poisson sampling scheme of (2.4.1) provided that

the three assumptions (Al), (A2), and (A3) hold.

THEOREM 2.4.1 If the model (2.3.4) satisfies assumptions (Al), (A2), and

(A3), then
i(P) = (M) and (P) = 7(M)

That is, the maximum likelihood estimators of / and are the same under

both sampling schemes-product-Poisson (2.4.1) and product-multinomial

(2.3.1).

Proof: Under the product Poisson assumption of (2.4.1) and (2.4.2), the

kernel of the log likelihood is

e(P)(; y) =y'(- e'1,

Therefore, letting 0 = vec(, A), the corresponding Lagrangian objective

function is

Q(0) = y'- e'1, + h'(\)A

and so to find the maximum (Poisson) likelihood estimator ^(P) = (i(P), 5(P))

we must solve the system of equations

9Q(O) y ei') + H(i())() (2.4.3)
o h( (P))0. (

The conclusion of the theorem now follows, since the equations (2.3.8) of






- 70 -


Theorem 2.3.1 and (2.4.3) yield exactly the same solutions and


(P) = (X'X)- X'Clog(Ae'P)) = (X'X)-lX'Clog(AeM') = (M)




As a corollary to Theorem 2.4.1 we have

COROLLARY 2.4.1 Provided the assumptions of Theorem 2.4.1 hold, the

estimated undetermined multipliers are invariant with respect to sampling

scheme, i.e.

A(M) = (P)




Proof: The proof follows immediately upon noting that equations (2.3.8)

and (2.4.3) yield exactly the same solutions. 0

A remark is in order. Basically, Theorem 2.4.1 enables us to conclude

that the sufficient and necessary condition of Birch (1963) holds. These

conditions are that the model be specified so that the Poisson ML estimators

necessarily satisfy the identifiability constraints that are required for the

multinomial model.

We now explore the asymptotic behavior of the (Poisson) ML estimator
b(P) = vec(i(P), i(P)). For the product-Poisson assumptions (2.4.1) and

(2.4.2), we can obtain the asymptotic distribution of 9(P) by formally replacing

the n, = min{ni} by p, = min{ei } and using the same arguments as those

used to derive the asymptotic distribution of i(M).

J0rgenson (1989) discusses limiting distributions for Poisson random

variables as the mean parameters, or equivalently /*, go to infinity. In this





- 71 -


case,

g(0o; Y) =( 0e)

has an asymptotic normal distribution with mean zero and asymptotic
covariance
(D(yo) 0)
0 0)
Using arguments similar to those used in the multinomial case, it follows that

(Y ) = (D(o) -OH) (O )+f ( OP+1
0 ) -H' ) jOp)

We conclude, as in the product multinomial case, that (P) Oo has an
asymptotic normal distribution with mean zero and asymptotic covariance

(D(po) -H)- (D(o) 0 D( C) -H r-
-H' 0 0 Oj 0\ -H' 0

But, this can again be simplified as it was in the multinomial case. It can be
shown that the asymptotic covariance can be rewritten as

D-1 D-'H(H'D-'H)-'H'D-1 0 (2.4.4)
0 (H'D-'H)-l ( 4)

where D = D(o) = D(e~O) and H = H(o).
Comparison of the Asymptotic Distributions. Provided assumptions (Al),
(A2), and (A3) hold, both (P) 0o and b(M) Oo have asymptotic normal
distributions with zero means and respective covariances given in (2.4.4) and
(2.3.12). Therefore, we have the following interesting results.
Result 1. The asymptotic covariances of (P) and i(M) are related by

var(()) = var(())- (2.4.5)
( 0 0)






72 -

Result 2. The asymptotic distributions of A(P) and ^(M) are identical and

it follows that the Lagrange multiplier statistic which has form


LM = A'(var(A))-l = A'(H'D-1H)A


is invariant with respect to the sampling scheme.

Result 3.
K (P) A (P)'
var((M)) = var(()) f (2.4.6)


Result 4.

var(/(M)) = var( ()) A (2.4.7)

where

A = (X'X)-X'C ) (e l )C'X(X'X)-.


and is nonnegative definite.

The notation var(.) used in these results denotes the asymptotic variance.

This is important since the finite sample variances may not even exist.

The proofs for Results 3 and 4 are straightforward. Basically, they

involve using the delta method and equation (2.4.5). The interested reader

will find an outline of the proofs in Appendix A.

In practice, it is of particular interest to evaluate the matrix A of equation

(2.4.7). Often, for convenience, the models are fit assuming the vector Y

is product Poisson and then inferences based on the maximum likelihood

estimates are made assuming that they are invariant with respect to the

sampling assumption. Birch (1963) and Palmgren (1981) derive rules for






73-

when these inferences, based on the two different sampling assumptions, will

be equivalent. However, they assume that the model is of a simple loglinear

form. That is, the Poisson model is assumed to have form



Ox = { R': = XI3}.



We will use the results of this section to derive more general rules for when

the two inferences will be equal. As a special case of these results, we will

arrive at the Birch and Palmgren results.

The following lemma will enable us to rewrite A of (2.4.7) in still a

simpler form.

LEMMA 2.4.1 Let Z = [Z1,..., ZK] be an r x K matrix of full rank K.

Suppose that X = [Xx,..., Xp] is an r x p (r > p > K) matrix of full rank p

such that M(X) M(Z), i.e. the range space of X contains the range space

of Z. Denote the T (K < T < p) columns of X that span a space that contains

M(Z) by {X,,,...,X,,}. Without loss of generality, suppose that the set of

vectors {XX,,...,Xv,} is a minimal spanning subset, i.e. the spanning set

of any r < T of these vectors does not contain the range space of Z. We

conclude that


3W e RTxK 9 (X'X)-'X'Z = JW,



where the p x T matrix J = [ex,...,e v] and ey, is the p x 1 vector

(0,...,0,1,0,...,0)' with the '1' in the vth position.





74 -

Proof: Let X. = [X,,...,XT]. Now, by assumption, M(X.) D M(Z).

Hence, there must exist a matrix W E RTxK 3 Z = X.W. Therefore,

(X'X)-1X'Z = (X'X)-'X'X.w = (X'X)-'(X'X.)w = JW

where J = (X'X)-'(X'X.) is as stated in the conclusion of the lemma. *

Before stating the next important theorem, let us write A in another

way. Assuming that (Al) holds, A can be written as

A-(Al A12)
A = A21 A22) (2.4.8)

where

A'J = (XiX) xj-C( )(D )CXj(XjXj)-

Now, if Ci is a contrast matrix, by assumption (A2), we can write


(X X)-'X 1( i m ) = J(iW('), (2.4.9)

where J(i) can arbitrarily be chosen to be equal to Xj and so W() = 0. On

the other hand, if Ci = IK then we have by (A3) that M(Xi) 2 M((lm,).

Therefore, we can invoke the result of Lemma 2.4.1 by setting Z = e .

Since M(Xi) 2 M(lm,,) = M(Z), the conditions for the lemma are satisfied.

Let Xi, = [X. (,),..., X ,)] be the miK x Ti (K < T, < pi) submatrix of X,

that has columns that form a minimal spanning subset for M(Z) = M(E-i-).

By Lemma 2.4.1,

3W() e RTxK 3 (X'XiX -1X, m) = J(')W(i). (2.4.10)

Here, J(i) = [e (),..., e ()], where the Ti elementary vectors correspond to the
1 Ti
columns {X. (i,..., X. )} of Xi that form a minimal spanning subset for the
ITi






75 -
range space of lm,, i.e. the Ti columns span a set that contains the range

space of elm, and any smaller set of columns will not span a set containing

the range space of ,lm, .

It follows that the matrices Aij of (2.4.8) can be written as


Ai' = J()W(W'(W()J,(j) (2.4.11)


where
j( [e) ,...,e()], if Ci = IqiK
W vzr(2.4.12)
Xi[, otherwise
and

W(i) W(), if C = IqK (2.4.13)
10, otherwise.
We now state a theorem of substantive importance.

THEOREM 2.4.2 Suppose that assumptions (Al), (A2), and (A3) hold. For
r = 1,2, if Cr is the identity matrix then let {v(r),... )} be the set of

indices that index those columns of X, that form a minimal spanning subset

for M($lm,). Then it follows that the relationship between the asymptotic

variances of the two estimators 3(M) and p(P) is


var( ((M) = var(/(")) (- 21 A1

where the pi x pj matrix Aii is a zero matrix whenever at least one of Ci or

Ci is a contrast or zero matrix. Otherwise, if both Ci and Cj are identity

matrices then


= 0, if (k,l) I {v) (i)*T" {v/1 (i)






- 76-


Proof: Since (Al), (A2), and (A3) hold, we can rewrite AiJ as in (2.4.11).

Now, if either C, or Cj are contrast or zero matrices, it is obvious by (2.4.9)

that Aij will have zero components, as stated in the theorem, since at least

one of W(i) or W(j) will be a zero matrix. On the other hand, if both C, and

Ci are identity matrices, then A'j can be rewritten as in (2.4.11) where

ji) = [e (, ...,e ],

J7i)= [(e ,,..., (,,

and the matrices W(i) and W(i) are elements of RTixK and RTjxK. Hence,



1 Ti e







where Wii = W(i)W'(i) is some Ti xT matrix. Now, since {e,} are elementary

vectors, we have that if


(k,1) {v ),..., )} x {vi),..., (),

then the component A' = 0. Otherwise, if (k, 1) is a member of this set, it

must be that A' is one of the elements of the matrix Wii. This completes

the proof.

The next two corollaries follow immediately from Theorem 2.4.1.

COROLLARY 2.4.2 If both C1 and C2 are contrast matrices then


var( (M)) = var((P )).






- 77 -


Proof: Since both 7C and C2 are contrast matrices it follows that W(1)

and W(2) are zero matrices. Therefore, the matrices Aij of the theorem are

zero matrices.

COROLLARY 2.4.3 Let C2 = 0,X2 = 0, and C1 A = I, = so that the model

(2.3.4) becomes


{h = {E R': = XP3, ee'( $1R) = n'},


i.e. a simple loglinear model with K subpopulations. Let {vl,..., vT} be the set

of indices that indez the columns of X that form a minimal spanning subset

for M(eflR). Then

var(A(M)) = var((P)) A,

where the elements of A are such that


Ak = 0, if (k, ) V {v,,..., VT}2



Proof: The proof is an immediate consequence of the theorem upon

identifying A" of the theorem with A of the corollary. The other matrices

A12, A21, and A22 will be zero since C2 = 0. 0

Corollary 2.4.3 is of practical importance and is essentially the result

shown by Palmgren (1981). In particular, if we parameterize the model in

such a way so that there is a parameter included for each of the K independent

multinomials (or K covariate levels), then the K columns of X corresponding

to these K 'fixed by design' parameters will form a basis (and hence a minimal

spanning subset) for AM(efRl). Therefore, if 3i and pj are not one of the






- 78 -


K parameters fixed by design, then cov(f(M), M)) = cov(1(P),~P). We

will illustrate the utility of the above results in the next chapter of this

dissertation.

The next section considers issues that may arise when computing the

model degrees of freedom. It also states some other miscellaneous results

with regard to the Lagrange multiplier statistic.



2.5 Miscellaneous Results

We begin this section by addressing practical issues that may arise during

nonstandard model fitting. Specifically, we will consider computing the model

and distance (or residual) degrees of freedom.

Computing model and distance degrees of freedom. Assuming the model

[Oh] of (2.3.5) is well defined, i.e. the u+I+K constraints are nonredundant,
we can compute the model degrees of freedom as in section 2.2. In that

section, we defined the model degrees of freedom as the number of model

parameters minus the number of independent constraints implied by the

model. Notice that in this application we have an additional 1 linear

constraints. The I constraints were not present in section 2.2. It follows

that the model degrees of freedom for [Oh] is


df[Oh] = s (u + 1 + K) (2.5.1)

where s is the number of cell means, u is the dimension of the null space of X',

1 is the number of linear constraints, and K is the number of identifiability

constraints.






79 -
To measure model goodness of fit, we can consider estimating some

hypothetical distance between model [Oh] and the saturated model (u = 1 = 0)

[O]. This distance, denoted S[eO; 0] has degrees of freedom

df ([Oh; 0]) = df[O] df[Oh]

= (s K) (s (u + + K)) (2.5.2)

= U + 1.
Notice that, had we considered the product Poisson model (2.4.2), the

distance degrees of freedom would be


df (6[O); O(P)]) = s ( ( + 1)) = + ,

which is identical to the product multinomial distance degrees of freedom of

(2.5.2).

We have assumed that the u +1+ K constraints are nonredundant, i.e.

each constraint is not implied by the other constraints. This may not always

be the case. To illustrate, consider the model specification for example 3 of

section 2.2.2. The model [OMH] implies that the two marginal distributions

are equal. We stated at the end of that example that the additional constraint

7r2+ 7r+2 = 0 was redundant. This can be seen since


72+ 7r+2 = r21 r12 = -(7rl+ 7r+i) = 0

That is, the constraints of model [OMH] imply that 7r2+ r+2 equals zero.

Had we blindly added this constraint, we may have incorrectly calculated

the model degrees of freedom as 1 and the distance degrees of freedom as 2.

Therefore, we must be very careful to have a set of nonredundant constraints

when computing degrees of freedom.






80 -

In practice, when models are more complicated, it may be difficult to as-

certain whether or not the model constraints are nonredundant. Fortunately,

there are two very useful results that help in this regard.

The first result is that when the constraints are redundant, the matrix

(H'D-1H) evaluated at some point in Oh is of less than full rank and is not

invertible. Therefore, in practice, if the algorithm (2.3.13) does not converge

due to G being singular, it may be due to redundant constraints, i.e. an ill-

defined model. The user should investigate and possibly respecify the model

should this occur. A caveat is that due to computational roundoff error, a

singularity may not occur even when the model is ill defined because the

iterate estimates, including the final estimate, may not strictly lie in Oh. The

next result may mitigate this problem.

A result that is useful in practice is that a necessary condition for the

constraints to be nonredundant or equivalently for the model to be well

defined, is that the Lagrange multiplier statistic be invariant to choice of

U, a matrix with columns spanning the null space of X'. Evidently, if the

user fits the model several times, each time using a different 'U' matrix, and

the Lagrange multiplier statistic varies (more so than can be explained by

roundoff error), then it must be that the model is ill defined.

Formally, this necessary condition can be stated as

THEOREM 2.5.1 Let U1 and U2 (U1 E U2) be any two full column rank

matrices satisfying ULX = 0, i = 1, 2. Denote the Lagrange multiplier statistic

evaluated using Ui by LM(Ui). If the matrix


H- Oh() _=- eA)CU,
Hi = 1=- Zt[log(e4'A')C'Uj, e4TL']
C~ ar





81-
is such that [Hi, ee] is of full column rank, i = 1, 2, and hence the models well
defined, then

LM(UI) = LM(U2),

i.e. the value of the Lagrange multiplier statistic is invariant with respect to
choice of U.
Proof: Denote the model specified in terms of Ui by [Oh,], i = 1,2. By
the definition of U, we know that the constraints implied by [Oh1] and [Oh,]
are equivalent. Hence, the solution to (2.3.8), or equivalently (2.4.3), under
either model is the same. Thus, in view of the first set of equations in (2.3.8),
any solution vec( A,) under model [Oh,] must satisfy


-(y-e) = Hi()A;, i= 1,2. (2.5.3)


Notice that since U1 f U2, we have that H () 5 H2() and by (2.5.3) Ai f A2.
Now, (2.5.3) implies that


Hi()Ai = Hg2()2. (2.5.4)


Also, since Hi() is assumed to be of full column rank, the variance of A,,


var(A,) = (H)()D-'(eZ)H,())-1 (2.5.5)


exists. Therefore, the Lagrange multiplier statistics LM(Ui), which have form


\[var(i)]-'5^, i= 1,2


(2.5.6)






82 -

exist. Finally, by (2.5.4)-(2.5.6), it follows that
LM(U1) = I[var(il)]-1l

= A'(H:(()D-l(e )Hz())Ai

= 2(H'( )D-1(ei)H2())\

= i'[var(l2)]-12

= LM(U2).
This completes the proof.

The final result of this section states that the Lagrange multiplier
statistic is exactly the same as the Pearson chi-squared statistic whenever the
random vector Y is product-Poisson or product-multinomial and the model

satisfies assumptions (Al), (A2), and (A3).

THEOREM 2.5.2 Assume that the product-multinomial model satisfies

assumptions (Al), (A2), and (A3). Let X2 denote the Pearson chi-squared

statistic, i.e.
2 = (y )'D-'()(y -i)

where A is the ML estimator under either of the sampling schemes-product-

multinomial or product-Poisson. It follows that the Lagrange multiplier

statistic LM is equivalent to X2. That is,

LM = X2.



Proof: By equations (2.5.3), (2.5.5), and (2.5.6) of the previous theorem's

proof and the fact that eM = M, we have that

LM = (y A)'D-'(f)(y 4) = X2


This is what we set out to show.






83-

2.6 Discussion



In this chapter, we discussed in some detail issues related to parametric

modeling. In particular, we followed the lead of Aitchison and Silvey (1958,

1960) and Silvey (1959) and described two ways of specifying models-using

constraint equations and using freedom equations. In section 2.2, distance

measures for quantifying how far apart two models are, relative to how close

they are to holding, were discussed. In particular, the power-divergence

measures (Read and Cressie, 1988) were used when the parameter spaces were

subsets of an (s 1)-dimensional simplex. Estimates of these distances were

developed based on very intuitive notions. Also, a geometric interpretation

of model and residual (or distance) degrees of freedom was given.

In section 2.3, we described a general class of multivariate polytomous

(categorical) response data models. The class of models, which satisfy

assumptions (Al), (A2), and (A3), were shown to satisfy the necessary and

sufficient conditions of Birch (1963) so that the models could be fitted using

either the product-Poisson or product-multinomial sampling assumption.

An ML fitting method was developed, using results of Aitchison and Sil-

vey (1958, 1960) and Haber (1985a, 1985b). The algorithm used Lagrangian

undetermined multipliers in conjunction with a modified Newton-Raphson

iterative scheme. The modification, which simplifies the method of Haber

(1985a), is to use a simpler matrix than the Hessian matrix. We replace

the Hessian matrix (of the Lagrangian objective function) by its dominant

part, which turns out to be easily inverted. Because the matrices used in the

algorithm proposed in this chapter are very large and must be inverted, this





84 -

modification is a very important one. A FORTRAN program 'mle.restraint'

has been written by the author to implement this modified algorithm.

The asymptotic behavior of the ML estimators computed under the two

sampling schemes-product-Poisson and product-multinomial-was investi-

gated. The method for deriving the asymptotic distributions represents a

modification to the technique of Aitchison and Silvey (1958). A comparison of

the limiting distributions of the two estimators was made in section 2.4. Some

very interesting results were obtained by studying the asymptotic behavior

in the constraint equation setting. In particular, Theorem 2.4.2 represents

a generalization of the results of Palmgren (1981). The theorem provides a

method for determining when the inferences about the freedom parameters

of a generalized loglinear model of the form C log Apl = X/f will be invariant

with respect to the sampling assumption. Palmgren (1981) developed some

similar results for the special case when the freedom parameters are part of

a loglinear model.

It is important to note that the asymptotic results are only valid if

the number of populations K is considered fixed and the expected counts

all get large at approximately the same rate. In particular, the asymptotic

arguments do not hold when the covariates are continuous, since the number

of populations (levels of the covariates) can theoretically run off to infinity.

The reason the arguments do not hold is that when we use the method of

Aitchison and Silvey (1958) it is required that the vector n,' lY;Y~ converge

in probability to zero as the total number of observations gets large. This is

the case only when n* = min{nl,..., nK} goes to infinity. This drawback






85 -

could prove to be temporary. It seems reasonable to assume in many cases,

that as long as the 'information' about each parameter is increasing without

bound, the estimators will be consistent and asymptotically normally dis-

tributed. For example, consider the logistic regression model with continuous

covariates. Although the nk's may all be 1, the ML estimators of the

regression parameters are often consistent and asymptotically normal.

Section 2.5 outlines some miscellaneous results. One result that is

important to the practicing statistician, is that the Lagrange multiplier

statistic is shown to be invariant with respect to choice of the matrix U

(of U'Clog Ay = 0) as long as the model is well defined. An important

implication of this result is that if one fits the model several times, each

time using a different 'U' matrix, and the Lagrange multiplier statistics

vary more so than can be explained by roundoff, then it could be that the

model is not well defined. Another interesting result is that the Lagrange

multiplier statistic is simply the Pearson chi-squared statistic X2 whenever

the assumptions (Al), (A2), and (A3) are satisfied.

Theoretically the ML fitting algorithm will work for any size problem.

Practically, however, the algorithm is certainly not a model fitting panacea.

The number of parameters that must be estimated gets very large, very fast.

Consider the case where 7 raters rate the same set of objects on a 5 point

scale. Even without covariates, the number of cell probabilities that must be

estimated is 57 = 78, 125. It seems the ML fitting method developed in this

chapter is, at least for now, useful for moderate size problems only. It can be

used to analyze longitudinal categorical response data when the number






86 -

of measurements taken on each subject is somewhere in the neighborhood of

2 to 6. This is not to take away from the utility of this chapter's algorithm,

but rather to indicate its breadth of application. In time, with increasing

computer efficiency, much larger data sets may be fitted using this algorithm.













CHAPTER 3
SIMULTANEOUSLY MODELING THE JOINT AND MARGINAL
DISTRIBUTIONS OF MULTIVARIATE POLYTOMOUS
RESPONSE VECTORS


3.1 Introduction

Often times, when given an opportunity to analyze multivariate response

data, the investigator may wish to describe both the joint and marginal

distributions simultaneously. We consider a broad class of models which

imply structure on both the joint and marginal distributions of multivariate

polytomous response vectors. To illustrate the need for such models, we

consider several settings where these models would be useful. For example,

when the multivariate responses represent repeated measures of the same

categorical response across time, one may be interested in how the marginal

distributions are changing across time and how strongly the responses are

associated. The simultaneous investigation of both joint and marginal

distributions is not restricted to the longitudinal data setting. Other examples

include the analysis of rater agreement, cross-over, and social mobility data.

The common thread tying all of these data types together is that the sampling

scheme is such that the different responses are correlated. In longitudinal

studies the same subject responds on several occasions. In rater agreement

studies, raters rate the same objects. In two-period cross-over studies, one

group of subjects receive the two treatments in one order and the other group

receive them in the other order. In social mobility studies, the socio-economic

87-






88 -

status of a father-son pair is recorded. When the responses are positively

correlated, these designs result in increased power for detecting differences

between the marginal distributions (Laird, 1991; Zeger, 1988).

This chapter considers the modeling of multivariate categorical responses

in which the same response scale is used for each response. The classes

of models used in this chapter are of the form considered in Chapter 2 of

this dissertation and hence are readily fit using the ML methods of that

chapter. In section 3.2, we give several examples that may be analyzed by

simultaneously modeling the joint and marginal distributions. We introduce

the classes of simultaneous Joint-Marginal models in section 3.3. Several

models are fitted to the data sets of section 3.2.


3.2 Product-Multinomial Sampling Model

Initially, we assume that a random sample of nk subjects is taken from

population k, k = 1,..., K. The number of populations, or covariate profiles,

K is considered to be some fixed integer. The subscript k is allowed to be

compound, i.e. the subscript k is allowed to represent a vector of subscripts

such as

k = (ki, k2,..., ik).


Suppose that there are T categorical responses V(1),..., V(T) of interest

and that each response is measured on the same response scale. Let

Vk = (Vk),..., VT))' be the random vector of responses for population k

and Vk,, u = 1,...,nk be the nk independent and identically distributed

copies of Vk, where Vk, denotes the response profile for the uth randomly





89 -

chosen person within population k. Notationally we have,

Vku ~ i.i.d. Vk, u = 1,...nk

For our purposes we can assume that each response takes on values in

{1,2,...,d} with probability one. Denote the probability that a randomly

selected subject from population k has response profile i = (i,..., iT)' by 7rk,

i.e.

P(Vk = (i,... iT)') =

where i e {1,...,d} x .-. x {1,...,d}.

The joint distribution of Vk = (V(),..., Vk(T))' is specified as {7rik}. The

marginal distributions of Vk will be denoted by {(i(t; k)}, t = 1,..., T, where

,(t; k)= P(V,(t) = i), i= 1i,...,d

Our objective is to model simultaneously the K joint distributions

{7TCk}, k= ,...,K

and the KT marginal distributions

{i(t; k)}, t= 1,...,T, k = l,...,K.

To help the reader better understand the notation, we consider the one

population bivariate case. When T = 2, the response profiles can be denoted

by i = (il,i2) = (i,j), where i = 1,...,d and j = 1,...,d. Since there is

just one population (or covariate profile) the subscript k is always 1 and is

therefore dropped. It follows that {7rij} is the joint distribution of (V('), V(2))'

and { i(t)}, t = 1, 2 are the two marginal distributions. That is,


7r, = P(V(I) = i, V(2) = j), i= l,...,d, j= l,...,d






- 90 -


and
(t) = '7i+ = P(V(I) = i), if t= l

7r+i = P(V(2) = i), if t = 2
for i= 1,2,..., d.

Now for each population k, consider the dT x 1 random vector of

indicators

'k = [I(V&=t1), .. ., I(V=_idT)]'

Notice that no information about the Vk is lost since 4k is a one-to-one

function of Vk. Also,


xk ~ ind. Mult(1, {7rik}), k= 1,..., K


Therefore, since we have randomly sampled nk subjects from each of the K

populations, we have that for given k


Tki k2, .,*k, ~ i.i.d. Mult(1, {7rik})


and hence the vector

nk
Yk = E k Mult(nk, {rik})
u=l

is sufficient for the family of distributions {rik} and {(i(t; k)}.

By independence across populations, the vector vec(,Y 2 ,.. .,YK) is

sufficient for the joint and marginal distributions of vec(V, V2,...,VK).

Further, the random vector vec(Yi, 2,...,YK) is product-multinomial, i.e.


Yk = (Yk,...,YR)' ~ ind Mult(n, {7ik}), k = 1,...,K


where 1,...,_R represent the R = dT different response profiles.






91-

Evidently, Yik represents the number of randomly selected subjects from

population k who have response profile i. That is, the {Yik} represent counts

resulting from a cross-classification of N = E'=1 nk subjects on T response

variables and a population variable. The data can be displayed in a dT x K

contingency table. By convention, we use lower case Roman letters to denote

realizations of random quantities. For example, yik represents a particular

realization of Yik.

Consider Table 3.1, taken from Hout et al. (1987).



Table 3.1. Interest in Political Campaigns

1960
Not Much Somewhat Very Much


Not Much


1956 Somewhat


Very Much


335


499


369


278 444 481 1203
Source: Hout et al. (1987), p. 166, Table 4




Each of 1203 randomly selected subjects was asked in 1956 how inter-

ested they were in the political campaigns. They responded on the 3-category

ordinal scale: 1 = Not Much, 2 = Somewhat, and 3 = Very Much.

Then, in 1960, each of the subjects was asked the same question and

responded on the same 3-category ordinal scale. Using the above notation,


155 116 64


91 237 171


32 91 246






92 -

we let V(1) and V(2) represent the responses in 1956 and 1960. Let yij, i,j =

1,2,3 represent the number of the N = 1203 subjects responding at level

i in 1956 and level j in 1960. Notice that there is just one population

of interest, we drop the population subscript altogether. Finally, for this

bivariate response example, the compound subscript i is replaced by ij. Table

3.1 summarizes the bivariate responses.

As another example, consider the cross-over data of Ezzet and White-

head (1991).


Table 3.2. Cross-over Data
B B
1 2 3 4 1 2 3 4


1 59 35 3 2 1
A 2 11 27 2 1 A 2
3 0 0 0 0 3
4 1 1 0 0 4


63 40 7 2
13 15 2 0
0 0 1 1
0 0 0 0


AB Sequence BA Sequence
(Group 1) (Group 2)


The counts displayed in Table 3.2 are from a study conducted by 3M

Health Care Ltd. to compare the suitability of two inhalation devices (A and

B) in patients who are currently using a standard inhaler device delivering

salbutomal. Two independent groups of subjects participated. Group 1 used

device A for a week followed by device B (sequence AB). Group 2 used the

devices in reverse order (sequence BA).

The response variables V(') (device A) and V(2) (device B) are ordinal

polytomous. Specifically, they are the self-assessment on clarity of leaflet

instructions accompanying the two devices, recorded on the ordinal four point

scale,






- 93 -


1 = Easy
2 = Only clear after rereading
3 = Not very clear
4 = Confusing.

For this example there are two populations of interest-Group 1 and

Group 2. Let yik represent the number of the nk subjects responding at level

i for device A and level j for device B, where nl = 142 and n2 = 144. Again,

the bivariate response profiles can be denoted by i = ij where i, j = 1, 2, 3, 4.

The bivariate responses are summarized in Table 3.2.


3.3 Joint and Marginal Models

Two types of questions that can be posed about Table 3.1 lead to quite

distinct types of models. One question is whether the interest in the political

campaigns was different at the two times. For example, the researcher

may wish to test the hypothesis that there was more interest in the 1960

political campaign than the 1956 political campaign. An investigation into the

marginal distributions is needed to test this hypothesis. For these bivariate

response data, the marginal distributions correspond to the row and column

distributions of Table 3.1. A second question that may be asked is whether

the two responses are associated and if so, how strong is the association. To

answer these questions, we must describe the dependence displayed in the

joint distribution of Table 3.1.

The marginal models we consider will be used to investigate whether

the probability that a randomly selected subject responds at level i or lower

in 1956 is different from the probability that a randomly selected subject

responds at level i or lower in 1960. In this sense, the comparison of marginal




Full Text
ON MODEL FITTING FOR MULTIVARIATE POLYTOMOUS
RESPONSE DATA
By
JOSEPH B. LANG
A DISSERTATION PRESENTED TO THE GRADUATE SCHOOL
OF THE UNIVERSITY OF FLORIDA IN PARTIAL FULFILLMENT
OF THE REQUIREMENTS FOR THE DEGREE OF
DOCTOR OF PHILOSOPHY
UNIVERSITY OF FLORIDA
1992
UNIVERSITY OF FL0R1BA LIBRARIES

ACKNOWLEDGMENTS
I would like to express my appreciation to Dr. Alan Agresti for serving
as my dissertation advisor. For the many comments, ideas, and lessons he
has shared with me, I am greatly indebted. Through his advisement and
guidance, he has taught me to appreciate and respect good statistical research
and teaching. He is a mentor worthy of emulation. I also want to express
my gratitude to Dr. Jane Pendergast, who also served on my dissertation
committee. I learned a great deal from her during the two years that I worked
in the Biostatistics Department. To all of the faculty at the University of
Florida, I extend my thanks. The statistics department, with its scholarly
and friendly atmosphere, proved to be a wonderful place to learn.
The influences of persons from my past are not forgotten. Without
Patrick Kearin’s stimulating teaching of high school math, I may never have
become interested in this subject. The genuine excitement delivered by Dr.
James Kepner, in his teaching of undergraduate statistics, was the reason I
decided to pursue an advanced degree in statistics.
I would like to thank my parents and the rest of my family for all of the
support and encouragement they have given over the course of my studies
and research. My friends and student colleagues deserve many thanks as
well. Finally, I would like to thank Kendra Paar for always being there to
support and encourage me while I was writing this paper.
n

TABLE OF CONTENTS
page
ACKNOWLEDGMENTS ii
LIST OF TABLES v
ABSTRACT vi
CHAPTERS
1 INTRODUCTION 1
1.1 A Brief Introduction to the Problem 1
1.2 Outline of Existing Methodologies—No Missing Data 3
1.3 Outline of Existing Methodologies—Missing Data 12
1.4 Format of Dissertation 14
2 RESTRICTED MAXIMUM LIKELIHOOD FOR A
GENERAL CLASS OF MODELS FOR
POLYTOMOUS RESPONSE DATA 17
2.1 Introduction 17
2.2 Parametric Modeling—An Overview 24
2.2.1 Model Specification 25
2.2.2 Measuring Model Goodness of Fit 33
2.3 Multivariate Polytomous Response Model Fitting 43
2.3.1 A General Multinomial Response Model 44
2.3.2 Maximum Likelihood Estimation 48
2.3.3 Asymptotic Distribution of Product-Multinomial
ML Estimator 56
2.3.4 Lagrange’s Method—The Algorithm 60
2.4 Comparison of Product-Multinomial and
Product-Poisson Estimators 67
2.5 Miscellaneous Results 78
2.6 Discussion 83
iii

page
3 SIMULTANEOUSLY MODELING THE JOINT AND
MARGINAL DISTRIBUTIONS OF MULTIVARIATE
POLYTOMOUS RESPONSE VECTORS 87
3.1 Introduction 87
3.2 Product-Multinomial Sampling Model 88
3.3 Joint and Marginal Models 93
3.4 Numerical Examples 98
3.5 Product-Multinomial Versus Product-Poisson
Estimators: An Example Ill
3.6 Well-Defined Models and the Computation of
Residual Degrees of Freedom 121
3.7 Discussion 132
4 LOGLINEAR MODEL FITTING WITH
INCOMPLETE DATA 135
4.1 Introduction 135
4.2 Review of the EM Algorithm 137
4.2.1 General Results 138
4.2.2 Exponential Family Results 140
4.3 Loglinear Model Fitting with Incomplete Data 144
4.3.1 The EM Algorithm for Poisson Loglinear Models 145
4.3.2 Obtaining the Observed Information Matrix 148
4.3.3 Inferences for Multinomial Loglinear Models 152
4.4 Latent Class Model Fitting—An Application 160
4.5 Modified EM/Newton-Raphson Algorithm 166
4.6 Discussion 170
APPENDICES
A CALCULATIONS FOR CHAPTER 2 172
B CALCULATIONS FOR CHAPTER 4 176
BIBLIOGRAPHY 193
BIOGRAPHICAL SKETCH 200
IV

LIST OF TABLES
page
2.1 Opinion Poll Data Configuration 22
3.1 Interest in Political Campaigns 91
3.2 Cross-Over Data 92
3.3 Joint Distribution Models—Goodness of Fit 100
3.4 Marginal Distribution Models—Goodness of Fit 101
3.5 Candidate Models in J(L x L + D) n M{U)—Goodness of Fit... 102
3.6 Estimates of Freedom Parameters for
Model J(L x L 4- D) n M(CU) 103
3.7 Freedom Parameter Estimates and Standard Errors 105
3.8 Estimated Cell Means and Standard Errors 106
3.9 Cross-Over Data Models—Goodness of Fit 110
3.10 Freedom Parameter ML Estimates for Model J(UÁ) n M(U).... 110
3.11 Children’s Respiratory Illness Data 112
3.12 Product-Multinomial versus Product-Poisson Freedom
Parameter Estimation 117
4.1 Observed cross-classification of 216 respondents
with respect to whether the tend toward
universalistic (l) or particularistic (2) values
in four situations (A,B,C,D) of role conflict 162
4.2 Parameter and Standard Error Estimates 164
4.3 Classification Probability Estimates 165
v

Abstract of Dissertation Presented to the Graduate School
of the University of Florida in Partial Fulfillment of the
Requirements for the Degree of Doctor of Philosophy
ON MODEL FITTING FOR MULTIVARIATE POLYTOMOUS
RESPONSE DATA
By
Joseph B. Lang
May, 1992
Chairman: Dr. Alan Agresti
Major Department: Statistics
A broad class of models that imply structure on both the joint
and marginal distributions of multivariate categorical (ordinal or nominal)
responses is introduced. These parsimonious models can be used to si¬
multaneously describe the marginal distributions of the responses and the
association structure among the responses. As a special case, this class
of models includes classical log- and logit-linear models. In this sense,
we address model fitting for multivariate polytomous response data from
a very general perspective. Simultaneous models for joint and marginal
distributions are useful in a variety of applications, including longitudinal
studies and studies dealing with social mobility and inter-rater agreement.
We outline a maximum likelihood fitting algorithm that can be used for
fitting a large class of models that includes the class of simultaneous models.
The algorithm uses Lagrange’s method of undetermined multipliers and a
modified Newton-Raphson iterative scheme. We also discuss goodness-of-fit
tests and model-based inferences. Inferences for certain model parameters
are shown to be equivalent for product-Poisson and product-multinomial
vi

sampling assumptions. This useful equivalence result generalizes existing
results. The models and fitting method are illustrated for several applications.
Missing data are often a problem for multivariate response data. We
consider inferences about loglinear models for which only certain disjoint
sums of the data are observable. We derive an explicit formula for the
observed information matrix associated with the loglinear parameters that
is intuitively appealing and simple to evaluate. The observed information
matrix can be evaluated at the maximum likelihood estimates and inverted
to obtain an estimate of the precision of the loglinear parameter estimates.
The EM-algorithm can be used to fit these incomplete data loglinear models.
We describe this algorithm in some detail, paying special attention to the
Poisson loglinear model fitting case. Alternative fitting algorithms are also
outlined. One proposed alternative uses both the EM and Newton-Raphson
algorithm, thereby resulting in a faster, more stable, algorithm. We illustrate
the utility of these results using latent class model fitting.
Vll

CHAPTER 1
INTRODUCTION
1.1 A Brief Introduction to the Problem
There are many situations when multiple responses are observed for each
‘subject’ in a group, or several groups. Here ‘subject’ is generically used to
refer to a randomly chosen object that generates responses. The multiple
responses could represent repeated measurements taken on subjects over time
or occasions. They could be the ratings assigned by several judges that all
viewed and rated the same set of slides (here, the ‘subjects’ are the slides).
Or, perhaps, it may be that several distinct or noncommensurate responses
are recorded for each subject. These responses are often categorical—ordinal
or nominal—and inevitably interrelated. This dissertation addresses issues
related to modeling and model fitting for multivariate categorical (ordinal or
nominal) responses.
Models for multivariate categorical response data are usually developed
to answer questions about (i) the association structure among the multiple
responses or (ii) the behavior of the marginal distributions of the response
variables. Specifically, a typical question of the first type is, “How are the
responses interrelated and is this interrelationship the same across the levels
of the covariates?” A typical type ii question is, “How do the (marginal)
responses depend on the covariates or occasions?” Historically, many models
(e.g. log- and logit-linear models) have been developed for the primary
- 1 -

- 2 -
purpose of answering the type i questions. Many of these models can easily
be fitted using maximum likelihood (ML) methods. These models typically,
however, are not useful for answering the type ii questions (Cox, 1972).
Marginal models—those models used to answer type ii questions—are not
as well developed. One reason for this is that ML fitting of these marginal
models is more difficult. At present, the method of weighted least squares
(WLS) is used almost exclusively for fitting these models.
Suppose that we are interested in answering questions of both types
i and ii. Usually the questions are addressed using two different models, a
joint distribution model and a marginal model, and fitting them separately. It
seems reasonable to want a model that can be used to address simultaneously
both questions. That is, we would like a model that simultaneously implies
structure on both the joint and marginal distribution parameters. To date,
there has been very little work done on the development and fitting of these
simultaneous models.
Whenever multiple responses are observed it is inevitable that there will
be missing data. There are several ways to fit the Poisson loglinear model with
incomplete data. One popular method is to use the EM algorithm to find the
ML estimates of the loglinear parameters. One drawback to this algorithm
is that a precision estimate of the ML estimators is not produced as a by¬
product. Several numerical techniques have been developed to approximate
the observed information matrix, which, upon inversion, will act as the
precision estimate. However, it would be of some convenience to derive an
explicit formula for the observed information matrix, at least in some special
cases.

1.2 Outline of Existing Methodologies—No Missing Data
We begin our discussion by considering the case of no missing data.
There are many methods for analyzing multivariate categorical (ordinal or
nominal) response data. These methods usually involve fitting (separately)
models for the joint or the marginal distributions of the response vectors.
In rare instances, simultaneous models for both the joint and marginal
distributions are considered. Maximum likelihood fitting methods for the
joint distribution models are simple and described in almost every standard
text on categorical data analysis. The fitting of marginal models using
ML methods is more difficult. Maximum likelihood fitting of the marginal
homogeneity model was considered by Madansky (1963) and Lipsitz (1988).
The fitting of a more general class of marginal models was considered
by Haber (1985a). Finally, the fitting of simultaneous models using ML
methods has only been addressed in the bivariate response case. The fitting
technique becomes very complicated when there are more than two categorical
responses. To appreciate the complexity of extending the technique to
multivariate response data, see section 6.5 of McCullagh and Nelder (1989)
or perhaps Dale (1986). In contrast, the ML fitting method of Chapter 2 can
easily be used to fit many marginal and simultaneous models. In the next few
paragraphs, we briefly describe the existing methods for modeling and model
fitting for multivariate categorical response data.
Modeling Joint Distributions Separately. One common method for analyz¬
ing multivariate categorical responses is to model the joint distribution only.
These models, which include classical log- and logit-linear models for the

- 4 -
joint probabilities, are useful for describing the association structure among
the responses. The last 30 years have seen the development of these methods
for analyzing multivariate categorical responses (Haberman, 1979; Bishop et
al., 1975; Agresti, 1984, 1990). For specificity, consider the following panel
study: One hundred randomly selected subjects were asked how interested
they were in the political campaigns. They were to respond on the 3 point
ordinal scale, (l) Not Much, (2) Somewhat, and (3) Very Much. Then four
years later the same group of subjects was asked to respond on the same
scale to the same question. A separate investigation into the association
structure would enable us to answer questions of a conditional nature. For
example, we could estimate the probability of responding ‘Very Much’ on the
second occasion given that the response at the first occasion was ‘Not Much’.
The description of these ‘transitional’ probabilities, although very interesting,
may not be completely satisfactory. We may also be interested in addressing
questions with regard to the marginal distributions. Perhaps we would like
to answer the question, “Are the distributions of responses to the political
interest question the same for each occasion?” Laird (1991), in a nice review of
likelihood-based methods for longitudinal analysis, mentions that the utility
of classical log- and logit-linear models is restricted to two situations: (1)
modeling the dependence of a univariate response on a set of covariates and
(2) modeling the association structure between a set of multivariate responses.
These models place structure on the joint probabilities and so they are not
directly useful for studying the dependence of the marginal probabilities on
occasion and other covariates. This problem was pointed out by several
authors (Cox, 1972; Prentice, 1988; McCullagh and Nelder, 1989;

- 5 -
Liang et al., 1991). An advantage of these models is that they are simple to fit
using either WLS (Grizzle et al., 1969), ML (McCullagh and Nelder, 1989),
or iterative proportional fitting (Bishop et al., 1975) methods. There are
many standard statistical programs available for fitting these models (SAS,
SPSS2, BMDP, GLIM, GENSTAT).
Modeling Marginal Distributions Separately. A second approach to an¬
alyzing multivariate categorical responses is to model only the marginal
distributions and to ignore the joint distribution structure. Full likelihood
methods that consider only models for the marginal probabilities tacitly
assume a saturated model for the joint distribution. Therefore, the models
may be far from parsimonious. In the non-Gaussian response setting, there
is a distinction between these marginal models and the transitional (or
conditional) models of the previous paragraph. Marginal models describe the
occasion-specific distributions and the dependence of those distributions on
the covariates. Transitional or conditional models describe the distribution
of individual changes over occasions. Models for these transitions can be
represented as probability distributions for the future state ‘given’ the past
states. Questions regarding transition probabilities can only be investigated
with longitudinal data. On the other hand, questions regarding the marginal
probabilities could theoretically be answered using cross-sectional data,
provided the cohort (subject) effects were negligible. Panel studies resulting
in longitudinal data result in more powerful tests for significance of within
cluster factors, such as occasion effect. This follows because there is a reduced
cohort effect; we are using the same panel of subjects at each occasion. For

-6 -
further discussion about the distinction between marginal and transitional
models, see Ware et al. (1988), Laird (1991), and Zeger (1988).
We will briefly discuss existing methods for making inferences about
the marginal probabilities separately. We will group these methods into 5
categories: (l) nonmodel-based methods, (2) WLS methods, (3) ML methods,
(4) Semi-parametric methods, and (5) other methods.
Nonmodel-based methods can be used to derive test statistics used for
testing specific hypotheses regarding the marginal distributions. Examples
include the Cochran-Mantel-Haenszel (1950, 1959) statistic which can be used
for testing the hypothesis of marginal homogeneity (MH) (cf. White et ah,
1982), McNemar’s (1947) statistic which can be used for testing the equality of
two dependent proportions, and Madansky’s (1963) likelihood-ratio statistic
for MH. Madansky’s statistic is a difference in fit of the model of marginal
homogeneity to the fit of the unstructured (saturated) model (see also Lipsitz,
1988 and Lipsitz et al., 1990). Many other relevant test statistics, some of
which are generalizations or modifications of the aforementioned (cf. Mantel,
1963; White et al., 1982), exist. Cochran’s (1950) Q statistic and Darroch’s
(1981) Wald-type statistic are examples of other test statistics that can be
used to test for marginal homogeneity.
Presently, if one was to fit a marginal model, say a generalized loglinear
model of the form ClogAfi = Xfl, where ¡j, is the vector of expected counts
in the full contingency table, he or she would most likely use the WLS fitting
algorithm. Most statistical software that fits these generalized loglinear
models does so using WLS. There are some advantages to using WLS. It
is computationally simple. Second-order marginal information is all that is

- 7 -
needed. And, the estimates are asymptotically equivalent to ML estimates.
Some disadvantages are that covariates must be categorical, sampling zeroes
create problems, and estimates are sensitive when second-order marginal
counts are small. The WLS method for analyzing categorical data was
originally outlined by Grizzle, Starmer and Koch (1969). Subsequently,
marginal models for longitudinal categorical data, or more generally mul¬
tivariate categorical response data, have been introduced and fitted using the
WLS method (Koch et ah, 1977; Landis and Koch, 1979; Landis et ah, 1988;
Agresti, 1989).
Maximum likelihood fitting of marginal models is more difficult since
the model utilizes marginal probabilities, rather than joint probabilities to
which the likelihood refers. When the responses are correlated, as they
invariably are, the marginal counts do not follow a product-multinomial
distribution. The full-table likelihood must be maximized subject to the
constraint that the marginal probabilities satisfy the model. Haber (1985a)
considers fitting generalized loglinear models of the form Clog Ap. = X(3 using
Lagrange multipliers and an unmodified Newton-Raphson iterative scheme.
The algorithm becomes very difficult to implement for even moderately large
tables. This is primarily due to the difficulty of inverting the large Hessian
matrix of the Lagrangian objective function. In this dissertation we consider a
modified Newton-Raphson that uses a much simpler matrix than the Hessian.
The matrix is easily inverted even for relatively large tables. Haber (1985b)
considers the estimation of the parameters ¡3 in the special case Clog/x = X/3.
We will use a modification of the method of Aitchison and Silvey (1958, 1960)
and Silvey (1959) to investigate the asymptotic behavior of the estimators of

-8 -
(3 in the more general model ClogA/i = X/3, thereby extending the work of
Haber (1985b). Another relevant paper, Haber and Brown (1986), considers
ML fitting of a model for the expected counts /i that has loglinear and
linear constraints. One can test hypotheses about the marginal probabilities
by comparing the fit of relevant models. Haber (1985a, 1985b) and Haber
and Brown (1986) only consider fitting the marginal models separately. No
attempt has been made to simultaneously model the joint and marginal
distributions.
Semi-parametric methods such as quasi-likelihood (Wedderburn, 1974)
and a multivariate extension, generalized estimating equations (GEE), have
become popular in recent years. The work of Liang and Zeger (1986), which
advocated the use of these GEEs, has been extended to cover the multivariate
categorical response data setting (Prentice, 1988; Zhao and Prentice, 1991;
Stram et ah, 1988; Liang et ah, 1991). With these semi-parametric methods,
the likelihood is not completely specified. Instead, generalized estimating
equations are chosen so that, when the marginal model holds, even if the
association among the multiple responses is misspecified, the estimators are
consistent and asymptotically normally distributed. These estimators, used
in conjunction with a robust estimator of their covariance (Liang and Zeger,
1986; Zeger and Liang, 1986; White, 1980, 1981, 1982; Royall, 1986), result
in consistent inference about the effects of interest. When the responses are
truly independent, the estimating equations with correlation matrix taken to
be the identity matrix, are equivalent to the likelihood equations. The GEE
approach requires the specification of a ‘working’ association or correlation
matrix. Examples of working associations include those that imply all

- 9 -
pairwise associations (measured in terms of odds ratios) are the same and
that the higher order associations are negligible (Liang et ah, 1991).
A related approach is known as GEE2. The consistency of these esti¬
mators follows only if both the marginal model and the pairwise association
model are correctly specified. This approach is a second order extension
of the GEEs of Liang and Zeger (1986) which are now termed GEEl. It
is second order because the estimation of the marginal model parameters
and the pairwise association model parameters is considered simultaneously.
The focus of both approaches, GEEl and GEE2, is usually on modeling
the marginal distributions—investigating how the marginal distributions
depend on occasion and covariates. The association is considered a nuisance.
Presently, there are no tests for goodness-of-fit of these models and so the
investigation into how well both models fit can be done only at an empirical
level. The assumption that higher order effects are negligible may not be
tenable. Testing procedures to assess the validity of these assumptions have
yet to be developed. Also, in contrast to WLS and ML methods, which
require only that the missing data be ‘missing at random’ (MAR), the semi-
parametric approaches require the missing data to be ‘missing completely
at random’ (MCAR). The assumption that the missing data mechanism is
MCAR is a much stronger assumption than MAR (Little and Rubin, 1986).
Finally, there are many other approaches to analyzing the marginal
probability structure separately. There are random effects models, whereby
subject-specific random effects induce a correlation structure on the multiple
responses. The marginal approach—the full likelihood is obtained by
averaging across the random effects—is computationally difficult (Stiratelli

- 10 -
et al., 1984). An alternative is to condition on the sufficient statistics
for the subject effects and consider finding the estimates by maximizing
the conditional likelihood. For further details on these conditional and
unconditional methods see Rasch, 1961; Tjur, 1982; Agresti, 1991; Stiratelli
et ah, 1984; Conaway, 1989, 1990. As yet another alternative, Koch et al.
(1980) give a bibliography for relevant nonparametric methods for analyzing
repeated measures data. Agresti and Pendergast (1986) consider replacing
the actual observations by their within cluster rank and testing for marginal
homogeneity using the ordinary ANOVA statistic for repeated measures data.
A three-stage estimator for repeated measures studies with possibly missing
binary responses has been developed by Lipsitz et al. (1992). This approach
is very similar to a generalized least squares approach, but it has some of
the nice features of the GEE approaches. One of these nice features is that
the estimators and their variance estimates are consistent under very mild
assumptions. An extension of this method to the polytomous response case
has yet to be developed.
Simultaneous Investigation of Joint and Marginal Distributions. There
has been very little work done to investigate simultaneously the joint and
marginal distribution structure. In some ways GEE2 is an attempt to
describe both distributions. However, only the pairwise (not the joint)
association structure is modeled; the higher-order associations are considered
a nuisance. Tests comparing nested models have not been developed in this
semi-parametric setting. Full likelihood approaches have been addressed
by Dale (1986), McCullagh and Nelder (1989, Chapt. 6), and Becker and
Balagtas (1991). Dale models the joint distributions of bivariate ordered

-11 -
categorical responses by assuming that the log global odds ratios follow a
linear model. The marginal probabilities are assumed to follow a cumulative
logit model. McCullagh and Nelder consider simultaneously modeling the
joint and marginal probabilities of a bivariate dichotomous response (two
distinct responses) by assuming that the log odds-ratios follow a linear
model and that the marginal probabilities follow a logit-linear model. Their
example included age as a categorical covariate. Finally, Becker and Balagtas
consider models for two-period cross-over data. The bivariate dichotomous
response was the response to the two different treatments. Order of treatment
application was considered a covariate. They assumed that the two log odds
ratios followed a linear model and that the marginal probabilities satisfied a
loglinear model. Because it is the marginal probabilities and not the joint
probabilities that satisfy a loglinear model, Becker and Balagtas refer to the
model as log nonlinear.
The ML model fitting approach used by each of these authors involves
a reparameterization of the likelihood, which is a function of the joint
probabilities, in terms of the joint and marginal model parameters. The
reparameterization in the bivariate response case—the case each author
considered—is somewhat complicated especially for multi-level responses. To
make matters worse, the extension of this method to general multivariate
polytomous responses looks to be extremely difficult. If the repaparameter-
izations are made so that the full likelihood is expressible in terms of the
joint and marginal model parameters, the likelihood can be maximized using
a Newton-Raphson-type algorithm. Basically, one must solve for the root of
some nonlinear score equation. This maximization approach is very sensitive

- 12 -
to the starting value in that convergence to a local maximum is not likely
unless the starting estimate is very close to the actual maximum. Finding
reasonable starting values is not a simple task. Dale (1986) outlines a method,
specifically for the models considered in that paper, for finding a starting
estimate.
In this dissertation, we outline an ML fitting method that can easily be
used to fit a large class of simultaneous models, including those considered
by Dale, McCullagh and Nelder, and Becker and Balagtas. The approach
involves using Lagrange’s method of undetermined multipliers along with a
modified Newton-Raphson iterative scheme. For all of the models considered,
an initial estimate for the algorithm is the data counts themselves along with
a vector of zeroes corresponding to a first guess at the values of the Lagrange
multipliers. The convergence of the algorithm is quite stable. The extension
to multivariate polytomous response data is straightforward.
1.3 Outline of Existing Methodologies—Missing Data
Missing data is often an issue when the response is multivariate in nature.
Missing data can also occur in more hypothetical situations. Examples
include loglinear latent class models (Goodman, 1974; Haberman, 1988)
and linear mixed or random effects models (Laird et ah, 1987). In latent
class analyses, a latent variable, which is unobservable, is assumed to exist.
Mixed or random effects models posit the existence of some unobservable
random variables that affect the mean response. In this brief outline, we will
consider ML methods for model fitting when the data are not completely
observable. Little and Rubin (1986) provide a nice summary of methods

- 13 -
for model fitting with incomplete data. There are many ways to find the
maximum likelihood estimators when the data are not completely observable,
each method having its positive and negative features. We could work directly
with the incomplete-data likelihood, which is usually complicated relative to
the complete-data likelihood, and use a Newton-Raphson or Fisher-scoring
algorithm. Palmgren and Ekholm (1987) and Haberman (1988) use these
methods to obtain maximum likelihood estimates and their standard errors.
Alternatively, we could avoid the complicated likelihood altogether and use
the Expectation-Maximization algorithm (Dempster et al., 1977). Sundberg
(1976) discusses the properties of the EM algorithm when it is used to
fit models to data coming from the regular exponential family. The EM
algorithm is one of the more flexible ML fitting algorithms for missing data
situations. We will primarily focus on this method for fitting loglinear models
with incomplete data.
Although the EM algorithm is easily implemented to fit loglinear models
with incomplete data, the algorithm does not provide an estimate of precision
of the model parameter estimators. Meng and Rubin (1991) outline a
supplemental EM (SEM) algorithm, whereby, upon convergence of the EM
algorithm, the variance matrix for the model estimators is adjusted to account
for missing data. The adjustment is a function of the rate of convergence of
the EM algorithm, which in turn is a function of how much information
is missing. Meng and Rubin numerically estimate the rate of convergence,
thereby obtaining an estimate of precision that reflects missingness. Although
this approach should prove to be applicable in the general situation, it still
is desirable to derive an explicit formula for the variance matrix that reflects

-14-
missingness. Other authors (Meilijson, 1989; Louis, 1982) have discussed
methods for estimating precision of model estimators when the data are
incomplete and the EM algorithm is used. Meilijson’s method involves EM-
aided differentiation, which is essentially a numerical differentiation of the
score vector. The method relies on the assumption that the observed data
components are i.i.d. (identically and independently distributed). Louis
gives an analytic formula for the observed information matrix based on the
incomplete data. The computation of the observed information matrix based
on this formula is not straightforward and must be considered separately for
each special application.
1.4 Format of Dissertation
In Chapter 2, we develop a maximum likelihood method for fitting a large
class of models for multivariate categorical response data. This development
follows a general discussion about parametric modeling. Concepts such as
degrees of freedom and model distances (or goodness of fit) are described at
an intuitive level. We also describe and compare the asymptotic distributions
of freedom parameter estimators under product-multinomial and product-
Poisson sampling assumptions. Chapter 3 has more of an applied flavor.
We consider simultaneously modeling the joint and marginal distributions
of multivariate categorical response vectors. A broad class of simultaneous
models is introduced. The models can be fitted using the techniques of
Chapter 2. Several numerical examples are considered. Chapter 4 outlines the
ML fitting technique known as the EM algorithm. This algorithm is used to
fit models with incomplete data. Some advantages and disadvantages of using

- 15 -
the EM algorithm are addressed. The most important disadvantage is that
the algorithm does not provide, as a by-product, a precision estimate of the
ML estimators. We derive an explicit formula for the observed information
matrix for the Poisson loglinear model parameters when only disjoint sums of
the complete data are observable. An application to latent class modeling is
considered. We also propose an ML fitting algorithm that uses both EM and
Newton-Raphson steps. The modified algorithm should prove to have many
positive features.
In this dissertation, we do not distinguish typographically between
scalars, vectors, and matrices. Parameters and variables are treated as ob¬
jects, their dimensions either being explicitly stated or implied contextually.
By convention, functions that map scalars into scalars, when applied to
vectors, will be defined componentwise. For example, if /j, represents an n x 1
vector, then
log = (log/i1,log/i2,...,log/in)\
We frequently use abbreviations that are common in the statistical
literature. They include ML (Maximum Likelihood), WLS (Weighted
Least Squares), IWLS (Iterative (Re)Weighted Least Squares), and EM
(Expectation-Maximization).
The range (or column) space of an n x p matrix X is denoted by M(X)
and is defined as {/lx : /x = X(3, f3 e Rp}. The symbols ® and 0 are the
binary operators ‘direct product’ and ‘direct sum’. The direct (or Kronecker)
product is taken to be the right-hand product. That is,
A®B = {Abij}.

-16-
The direct sum, C, of two matrices A and B is defined as
C = A® B = 0).
The symbol D(n) represents a diagonal matrix with the elements of /¿ on the
diagonal. That is,
/>i 0 ... 0\
0 fJ.2 ••• 0
V 0 0 ... ¿¿n /
In Chapter 4, we make use of the bracket notation often used by
statistical and mathematical programming languages (e.g. Splus, Matlab).
To illustrate the notation, consider a matrix A. The (sub)matrix A[, -2] is
then matrix A with the second column deleted. Similarly, the matrix A[-3,]
is the matrix A with the third row deleted.
Equation numbering is consecutive within sections of a chapter, the
first number representing the chapter in which it appears. For example, the
thirteenth equation in section 2.3 is equation (2.3.13). Within each appendix,
the equations are numbered consecutively. For example, the third equation
in Appendix B is numbered (B.3). Tables are numbered consecutively within
chapters so that, for instance, Table 3.2 represents the second table within
Chapter 3. Theorems, lemmas, and corollaries are numbered independently
of each other. All are numbered consecutively within sections. Therefore,
Corollary 3.2.2 is the second corollary within section 3.2 and Theorem 2.3.1
is the first theorem within section 2.3.

CHAPTER 2
RESTRICTED MAXIMUM LIKELIHOOD FOR A GENERAL
CLASS OF MODELS FOR POLYTOMOUS RESPONSE DATA
2.1 Introduction
In this chapter, we consider using maximum likelihood methods to fit a
general class of parametric models for univariate or multivariate polytomous
response data. The models will be specified in terms of freedom equations
and/or constraint equations. These two ways of specifying models will be
discussed at length in section 2.2. The model specification equations may be
linear or nonlinear in the model parameters. Specifically, if represents the
s x 1 vector of expected cell means, the linear constraints will be of the form
L/j, = d and the nonlinear constraints will be of the form U'Clog^Afi) =
0. The freedom equations will have form Clog(A/i) = X(3, where the
components of the vector /3 are referred to as the freedom parameters. In
Chapter 3 of this dissertation, we discuss more specifically models that can
be specified in terms of these constraint and freedom equations. The models
of that chapter allow one to simultaneously model the joint and marginal
distributions of multivariate polytomous response vectors.
The maximum likelihood, model fitting algorithm of this chapter utilizes
Lagrange multipliers and a modified Newton-Raphson iterative scheme. In
particular, the models will be specified in terms of constraint equations and
the log likelihood will be maximized subject to the constraint equations being
- 17 -

- 18 -
satisfied. One common optimization algorithm found in the mathematics
literature is Lagrange’s method of undetermined multipliers. We show that
Lagrange’s method is easily implemented for ML fitting of the models under
consideration in this chapter. One problem with Lagrange’s method of
undetermined multipliers for ML fitting of statistical models has been that it
becomes computationally infeasible for large data sets. By using a modified
Newton-Raphson method which involves inverting a matrix of a simpler form
than the more complicated Hessian, we consider fitting models to relatively
large data sets.
We also explore the asymptotic behavior of the estimators within the
framework of constraint—rather than freedom—models. Usually, asymptotic
properties of model and freedom parameter estimators are studied within the
framework of freedom models. Aitchison and Silvey (1958, 1960) and Silvey
(1959) studied the asymptotic behavior of the model parameter estimators
when the model is specified in terms of constraint equations. Following the
arguments of Aitchison and Silvey, we derive the asymptotic distributions of
both the model and freedom parameter estimators.
Previous work by Haber (1985a) addressed maximum likelihood methods
for fitting models of the form
C\ag(Ati) = XP,
to categorical response data. Subsequently, Haber and Brown (1986)
discussed ML fitting for loglinear models that were also subject to the
linear constraints L/u, = d, where these constraints necessarily include the
identifiability constraint required of the vector of product-multinomial

- 19 -
cell means. Both of these papers advocated the use of Lagrange’s method
of undetermined multipliers to find the maximum likelihood estimates of
the model parameters /x. The method of Haber (1985a) involved using
the (unmodified) Newton-Raphson method which becomes computationally
unattractive as the number of components in ¡x gets moderately large. Both
Haber (1985a) and Haber and Brown (1986) were primarily concerned with
measuring model goodness of fit and therefore did not consider estimation
of freedom parameters. Haber (1985b) did consider estimation of freedom
parameters, but only when the simpler model C log/i = X/3 was used. One of
the several ways that we extend the work of Haber (1985a, 1985b) and Haber
and Brown (1986) is to consider estimation of the freedom parameters when
the more general model ClogAfx = X(3 is used.
Others have considered ML fitting of nonstandard models for multivari¬
ate polytomous response data. Laird (1991) outlines the different approaches
taken by different authors. As an example, Dale (1986) considered ML fitting
for a particular class of models for bivariate polytomous ordered response data
which were of the form
C\ log(Aifx) — Xi/3i, g(A2fx) — X2(32
Specifically, the first freedom equation specifies a loglinear model for the
association between the two responses measured by the global cross-ratios
(cross-product ratios of quadrant probabilities) so that C\ and A\ are of
a particular form. The second set of freedom equations specifies some
generalized linear model (McCullagh and Nelder, 1989) for the marginal
means or probabilities. Maximum likelihood estimators for the association

- 20 -
model freedom parameters 0i and the marginal model freedom parameters
02 were simultaneously computed by iteratively solving the score equations
via a quasi-Newton approach. To use this maximization technique, the score
functions, which involve the cell probabilities, must be written explicitly
as a function of the freedom parameter 0 = vec^, 02). A nontrivial
approach to finding reasonable starting values for 0 is discussed by Dale
(1986). Along with Dale, McCullagh and Nelder (section 6.5, 1989) and
Becker and Balagtas (1991) consider writing the score as an explicit function
of the freedom parameters so that the marginal and association freedom
parameter estimates may be computed simultaneously. In general, when there
are more than two responses, this is not a simple task and so an extension
of this method to multivariate polytomous response data models will be very
messy indeed. Also, convergence of the iterative scheme requires good initial
estimates of the freedom parameter 0. These may be very difficult to find. In
contrast, the maximization approach of this chapter, which is similar to Haber
(1985a) and Haber and Brown (1986), is shown to be easily implemented for
fitting multivariate polytomous response data models. With this technique,
it is not necessary to write the cell means as an explicit function of the
freedom parameters. Further, initial estimates of the freedom parameters,
which are difficult to find, are not needed for this technique. Instead, only
initial estimates of the cell means and undetermined multipliers are needed.
Reasonable initial estimates of the cell means are the cell counts themselves.
While a reasonable initial estimate of the vector of undetermined multipliers
is the zero vector—the value of the undetermined multipliers when the model
fits the data perfectly.

- 21 -
We will now introduce the class of models that we will consider for the
remainder of this chapter and the next, more applied chapter. The models
have form
logC-^iAO — XiPi, C2 log(A2/x) — AT2/32, L[i — d
where the linear constraints include the identifiability constraints. Later,
when we study the asymptotic behavior of the ML estimators, we will
require the components of d to be zero unless they correspond to an
identifiability constraint. These models, which are of the form C\og(Afi) =
X/3, Lfj, = d, will allow us to model both the joint and marginal distributions
simultaneously when dealing with multivariate response data. The bivariate
association model of Dale (1986) is a special case of these models, as we
can specify the matrices C\ and A1 so that Ci log(Ai/¿) is the vector of log
bivariate global cross-ratios. Restricting the marginal models to have form
C2 log(A2/¿) — -X"2/?2, rather than allowing the marginal means to follow a
generalized linear model, as Dale (1986) did, is not overly restrictive. In
fact, many of the generalized linear models for multinomial cell means can be
written in this form. For example, loglinear, multiple logit, and cumulative
logit models are of this form. Also, unlike Haber (1985a) and Haber and
Brown (1986), we will be concerned with estimation of the freedom parameter
¡3 = vec(/?i, /32), thereby allowing for model-based inference.
Model-based inferences usually refer to inferences based on freedom
parameters. With freedom equations, we have the luxury of choosing a
parameterization that results in the freedom parameters having meaningful
interpretations. For instance, a freedom parameter (3 may be chosen to

- 22 -
represent a departure from independence in the form of a log odds ratio.
More generally, we usually will try to parameterize in such a way so that
certain parameters will measure the magnitude of an effect of interest.
For example, consider an opinion poll where a group a subjects were
asked on two different occasions whether they would vote for the President
again in the next election. Suppose they were asked immediately after the
President took office and again after the President had served for two years.
The researcher may be interested in determining whether the distribution of
response changed from Time 1 to Time 2 and if so, assess the magnitude of
the change. The data configuration can be displayed as in Table 2.1.
Table 2.1. Opinion Poll Data Configuration
Data
Time 2
yes no
Probabilities
Time 2
yes no
Time 1 yes
2/n
2/12
Time 1 yes
*11
7Ti2
no
2/21
V22
no
7121
*"22
*"+l *"+2
*"l +
*"2+
We could formulate a model of the form Clog(Afi) = X(3 in such a way
so that the freedom parameter (3 has a nice interpretation with respect to the
hypothesis of interest. One such model is
log(|4g) = a + * = 1,2 (2.1.1)
where the parameter is a marginal probability, i.e.
if i = 1
if i = 2

- 23 -
and, for identifiability of the freedom parameters,
Pi = ~P2 = P-
Model (2.1.1) is a simple logit model for the marginal probabilities {7r¿+} and
{7rj }. The parameter p measures the magnitude of departure from marginal
homogeneity in that p = 0 if and only if there is marginal homogeneity.
One could use the Wald statistic p/se(p) to test the hypothesis. If the
null hypothesis is rejected, we can assess the magnitude of departure from
marginal homogeneity by computing a confidence interval for 2p which is the
log odds ratio comparing the odds that a randomly chosen subject responds
‘yes’ at Time 2 to the odds that a randomly chosen subject responds ‘yes’ at
Time 1.
This simple example illustrates the utility of using freedom parameters
and the corresponding model-based inferences. For this reason, this chapter
will be concerned with making inferences about both the model parameters
p and the freedom parameters /3.
The contents of the following sections are as follows. In section 2.2,
we provide an overview of parametric modeling. The two ways of specifying
models—via constraint equations and via freedom equations—are discussed
at length in section 2.2.1. It is shown that a model specified in terms of
freedom equations can be respecified in terms of constraint equations. In
particular, the freedom equation Clog(j4/i) = Xf3, which actually constrains
the function C\og(Ap) to lie in some manifold spanned by the columns of X,
is equivalent to the constraint equation U'Clog(Ap) = 0, where the columns
of U form a basis for the null space of X'. Other topics covered in section 2.2

-24-
include interpretation and calculation of ‘degrees of freedom’ and measuring
model goodness of fit.
We describe a general class of models for univariate or multivariate
polytomous response data in section 2.3.1. The data vector y is initially
assumed to be a realization of a product-multinomial random vector. We
describe the asymptotic behavior of the product-multinomial ML estimators
in section 2.3.3. Lagrange’s method of undetermined multipliers is used to
find restricted maximum likelihood estimates of the model parameters and
the freedom parameters. The actual algorithm is described in detail in section
2.3.4.
In section 2.4, we explore the relationship between the product-multinomial
and product-Poisson ML estimators. General results that allow one to
ascertain when inferences based on product-Poisson estimates are the same as
inferences based on product-multinomial estimates are shown to follow quite
directly when one works within the framework of constraint models. Theorem
2.4.2 of this section, represents a generalization of the results of Birch (1963)
and Palmgren (1981).
2.2 Parametric Modeling—An Overview
Inferences about the distribution of some n x 1 random vector Y are
often based solely on a particular realization y of Y. In parametric modeling
it is often the case that the distribution of Y is known up to an s x 1 vector
of model parameters 0; i.e. it is ‘known’ that
Y ~ F(y-9), 0 6 0,
(2.2.1)

- 25 -
where 0 is some (s-«^-dimensional (q > 0) subset of R3 known to contain the
true unknown parameter 9*. The cumulative distribution function F maps
points in Rn into the unit interval [0,1] and is assumed to be known.
In general, we will allow the dimension s of 0 to grow with n. For
example, let Y = (Yi,..., Yn) have independent components such that
Yi ~ ind G(yi]Zi(6)), t = l,...,n,
where Z{(9) is some function of 9 associated with the ith component of Y.
The function could be defined as z,(#) = 9i, in which case s = n. Or, on
the other hand, Z{ could be a mapping from Rs to R1 with s fixed.
2.2.1 Model Specification.
In parametric settings, models for the data, or more precisely, models for
the distribution of Y, can be completely specified by recording the family of
candidate distributions that F may belong to. That is, one must specify the
form for F(-]9) and the space Om that is assumed to contain the true value
9* of 9. In parametric modeling, the form of F1(-; 9) is assumed known, but
the true value 9* is not. Denote a parametric model by [F(-]9)-,9 £ 0m] or
more simply by [0m]- We say the model [0m] ‘holds’, if the true parameter
value 9* is a member of 0m, he.
[0m] holds 9* £ 0M-
A model does not hold if 9* g 0m-
The objective of model fitting is to find a simple, parsimonious model
that holds (or nearly holds). By parsimonious, we mean that the vector 9 can
be obtained as a function of relatively few unknown parameters. An example

- 26 -
of a parsimonious model for the distribution of an n-variate normal vector
with unknown mean vector fi and known covariance is [0/3], where
0/3 = {fj. e Rn : Hj = (3, j = 1,..., n, (3 unknown}.
Notice that all n components of /j, can be obtained as a function of
one unknown parameter /3. Thus, all of our estimation efforts can be
directed towards the estimation of the common mean (3. An example of a
nonparsimonious model is the so-called saturated model [0], where
0 = {/i : n g Rn} = Rn.
In this case, fi is a function of n unknown parameters.
The question of whether or not the parsimonious model holds is an
entirely different matter. Practically speaking, a model will rarely strictly
hold. Therefore, we will often say a model holds if it nearly holds, i.e. for
some small e
inf ||«* - #|| < e.
Om
Without delving too much into the philosophy of model fitting and the
simplicity principle (Foster and Martin, 1966), we point out that for a model
to be practically useful it must be robust to the ‘white noise’ of the process
generating Y. That is, it should account for only the obvious systematic
variation. A model would be said to be robust to the white noise variability,
if the model parameter estimates based on different realizations of Y are very
similar. As an example, if instead of [0^], the saturated model [0] was used
to draw inferences about the normal mean vector /i, we would find that the
model fit perfectly, but that upon repeated sampling the model estimates

-27-
would change dramatically. Thus, the model is not robust to the white noise
of the process. On the other hand, the parsimonious model [0^] estimates
would change very little from sample to sample, varying with the sample
mean of n observations. This model is robust to the white noise variability.
Therefore, if the model would hold, or nearly hold, we would say it was a
good model.
Freedom Models. In the previous n-variate normal example we specified a
model [0^] in terms of some unknown parameter ¡3. Aitchison and Silvey
(1958, 1960) and Silvey (1959) refer to the parameter [3 as a ‘freedom
parameter’ and the model [0^] as a ‘freedom model’. These labels are
reasonable since we can measure the amount of freedom we have for estimating
9 by noting the number of independent freedom parameters there are in the
model. The model [0^] has one degree of freedom for estimating the mean
vector /i. Thus, once an estimate of the single parameter (3 is obtained the
entire vector g, can be estimated; it is a function of the one parameter ¡3.
Notice that ‘degrees’ of freedom correspond to integer dimension in that a
degree of freedom is gained (lost) if we introduce (omit) one independent
freedom parameter thereby increasing (decreasing) the dimensionality of 0^
by one.
In general we will denote a freedom model by [0^], where
ex = {0eQ:g(9) = X(3i¡3eRr}
The function g is some differentiable vector valued function mapping 6 £ 0
into r-dimensional Euclidean space Rr. The ‘model’ matrix X is an r xp full
column rank matrix of known numbers. To calculate degrees of freedom for

- 28 -
[©x] we initially assume g satisfies
V#o e ©x5
(M)
V 06'
#0
is of full row rank r.
It also will be assumed that the constraints implied by g{6) = X/3 are
independent of the q constraints implied by the model [0] of 2.2.1. Well
defined models will satisfy these conditions. For example, any g that is
invertible satisfies the derivative condition. Actually this derivative condition
is not a necessary condition for the model to be well defined. Later, we will
show that g need only satisfy a milder derivative condition.
The degrees of freedom for the model [0x] can be obtained by subtract¬
ing the number of constraints implied by [©x] from the total number of model
parameters, s. The number of constraints implied by [©x] is (r - p) + q, the
dimension of the null space of X' plus the q constraints implied by model [0].
Hence, the model degrees of freedom for [0x] is
df[®x] = s-(r-p + q) (2.2.2)
In view of (2.2.2) the model degrees of freedom, an integer measure of freedom
one has for estimating 9, is an increasing function of p the number of freedom
parameters. In fact, for the special case when q = 0 and g(9) = 9 (so s = r),
we have that the number of degrees of freedom for model [0x] is simply p,
the number of freedom parameters. This gives us another good reason for
calling (3 a freedom parameter and [©x] a freedom model.
Constraint Models. Notice that
{9ee:g(e) = X(3,f3eRp}
(2.2.3)
can be rewritten as
{# e 0 : U'g(9) = 0},

- 29 -
where U is an r x (r -p) full column rank matrix satisfying U'X = 0, i.e. the
columns of U form a minimal spanning set, or basis, for the null space of X'.
Letting u = r - p and h*(8) = 0 be the q constraints implied by [0], we can
write the (u + g) x 1 vector of constraining functions as h(9) = [hi(6), h*(8)]'
where hi = U'g. We rewrite the freedom model [0x] of (2.2.3) as [0/,], where
Sh = {6 e R3 :h{8) = 0}. (2.2.4)
Aitchison and Silvey (1958,1960) refer to model [0^] as a constraint model.
Every freedom model can be written as a constraint model.
We present a few simple examples to illustrate the equivalence between
the two model formulations—freedom and constraint.
Example 1. Let Y¿ ~ ind N(/3,a2), i = 1 ,...,n, where cr2 is known.
This model can be specified as the freedom model [0_y], where
©x = {p E Rn '■ p = 1 n/3, ¡3 unknown }
or equivalently it can be expressed as the constraint model [0/,], where
Qh = {p g Rn : U'p = 0}
and U' is the (n - l) x n matrix
/I -1 0 0 0 \
u<= l 0 -10 ... 0
\1 0 0 0 ... -1/
It is easily seen that 0^ = 0/, and that the model degrees of freedom is
df[Qx] = n-(n - 1) = 1.
Example 2. Let Y¿ ~ ind N(pi = /30 + fliXi, cr2), i = 1 ,...,n, where is known. This model can be specified as the freedom model [0_x]5 where
Ox = {P e Rn : Pi =/5o +fiiXi, * = l,...,n}

- 30 -
or assuming that each is distinct, as the constraint model [©/,]> where
eh = {n G Rn : U'n = 0 }.
Here U' is the (n - 2) x n matrix
/
-l
i
U' =
+
-i
Z2—Zl Z2 —Xl X3 — X2 X3—X2
-1 1 1
0
x2 X\
\ X2-X-Í
Zj-Zl
Z4—Zs Z4—Zs
Z2-Zl
0
0
0
0
0
0
0 \
0
-1
—1 zr
-in-l /
Notice that U'fi = 0 implies that
Hj+i ~ _ Hk+i ~
xj+1 — xj xk+l ~~ xk
, Vk,j.
That is, the n means fall on a line. As before, it can be seen that ©x = ©&
and that the model degrees of freedom is df[Qh] = n — (n — 2) = 2.
Definitions. We will assume that the constraining function h satisfies
some reasonable conditions so that the model is well defined. We first present
some definitions.
(1) A model [©/,] is said to be ‘consistent’ if Qh 7^ 0.
(2) A consistent model [©/,] is said to be ‘well-defined’ if the Jacobian
matrix for h is of full row rank v — u -f q at every point in Qh. That is,
v*“e e*-
#0
is of full row rank v.
(3)A model [©/,] is said to be ‘ill-defined’ if it is not well-defined, i.e.
3i fe (ahW
*,£°b
00
is not of full row rank u.

- 31 -
(4) An ill-defined model [0/,] is said to be ‘inconsistent’ or ‘incompatible’
if ©/, = 0.
Briefly, any reasonable model will have a nonempty parameter space and
hence will be consistent. The Jacobian condition of definition (2) is similar
to the condition required in the Implicit Function Theorem (see Bartle, 1976).
Basically, this condition requires the constraints to be nonredundant so that,
at least theoretically, the constraint equations can be written uniquely as
a function of a smaller set of parameters. An ill-defined model has been
specified with a redundant set of constraint equations. Using the lingo of
the optimization literature, two constraints are redundant if, for each point
in the parameter space, both of the constraints are ‘active’ or both of the
constraints are ‘inactive’. That is, for all parameter values, if one constraint
is active (inactive) then the other is necessarily active (inactive).
It should be noted that the above definitions are in terms of the
constraint formulation of a model. This is sufficient since freedom models can
be written as constraint models. For convenience, we give sufficient conditions
for a freedom model to be well-defined.
A consistent freedom model is well-defined if it satisfies the following two
conditions:
(i) The constraints implied by g{6) = X/3 are independent of the q
constraints implied by [0].
(ii) The Jacobian matrix of g evaluated at any point in [0x] is of full row
rank r, i.e.
ggffl
d0‘
00
)•
is of full row rank r.
V#o 6 0x>

- 32 -
The sufficiency of conditions (i) and (ii) can be seen by observing that
(ii) implies that hi = U'g has a full row rank Jacobian since U' is of full row
rank and (i) implies that h = (hi,h*)' has full row rank Jacobian. These
sufficient conditions are by no means necessary for a model to be well defined
as the Jacobian of h may be of full row rank u even when the Jacobian of g
is not of full row rank.
Notice that the model matrix has nothing to do with whether or not a
model is well defined. In particular, one may think that the model [0^] is
ill-defined whenever the r x p matrix X is not of full column rank; i.e. the
freedom parameters are nonestimable. However, the model can be rewritten
as a constraint model with the full column rank matrix U spanning the null
space of X, which has dimension less than p - r. It follows that if g satisfies
(i) and (ii), then the model [©x] will be well-defined. The only reason we
have taken X to be of full column rank is to avoid using generalized inverses
when working with the freedom parameters.
To illustrate the use of these definitions, we consider the model [0M],
where
Qm = {9 e Rn : Md - d = 0}.
The model will be well defined if dh/d6' = M is of full row rank. It is
inconsistent if the linear system of equations M9 = d is inconsistent.
If a model [0/,] is well defined, then the constraints implied by the model
are all independent in that no constraint can be implied by the others. We
will consider only well-defined models when calculating degrees of freedom.

- 33 -
As before, we calculate degrees of freedom for a model as the difference
between the number of model parameters s and the number of independent
constraints v implied by the model, i.e.
df[Qh] = s-(r-p + q) = s-(u + q) = s- u
Notice that for the constraint model, model degrees of freedom is a decreasing
function of the number of independent constraints u.
Finally, it should be noted that models may be specified in terms of
both freedom equations and constraint equations. In fact, in subsequent
sections this will be the case. However, without loss of generality, we will
concentrate on constraint models since any model can be written in the form
of a constraint model.
2.2.2 Measuring Model Goodness of Fit
Inferences about model parameters are reliable only if the model is
‘good’. A good model should be well defined (or at least consistent). It
should be simple and parsimonious. Finally, the model should be relatively
close to holding.
To assess whether or not the model holds, we will need the concept of a
distance between two models. To begin, we will assume there is some measure
of distance between two hierarchical parametric models. (Two models [0i]
and [@2] are hierarchical if 02 C 0! and d/[02] < d/[0i] whenever 0j ^ 02.)
This (parametric) distance will be a quantitative comparison of how close
the two models are to holding. Thus, if both models hold the distance is
zero. The distance will also be independent of the model degrees of freedom.

- 34 -
Recall that the form of F(.] 9) is assumed known. Therefore, the distance will
measure how far the true parameter is from falling in the parametric model
space. Suppose, firstly, that ©! and @2 are general parameter spaces. That
is, 6 g ©i u ©2 does not necessarily define a probability distribution. In other
words, 9 need not fall in a subset of an (s- l)-dimensional simplex. Let a(9)
and b(9) be vector or matrix valued functions of the unknown parameter 9.
Define a distance between two hierarchical models [©1] and [02] (©2 C ©j) as
i[02; e,] = inf ||6(0)(a(e) - a(0’))||* - inf ||6(»)(o(«) - a(0*))||2.
Notice that a and b can be chosen so that
(1) 6[02;©i]>O
(2) <$[02;©i] = 0, iff ©! and 02 hold.
For example, consider the case Y ~ MVNn(fi,cr2In). Suppose that
[©] = {(/b*2) e Rn, 0}
[©i] = {(m, ct2) : // = //0, 0}
[02] = {(aí, 0}
[©3] = {(/b ^2) •• A4 - ln<*, OCER,a2> 0}.
In this example, each component of Y has a common variance a2. It seems
reasonable that differences between any ¡J,j and the true mean ¡x*- are equally
important. Hence, a natural distance between any two of these models is
= inf ||//-/i*||2 - inf ||//-/z*||2.
WM2 {^)M1
Notice that a(/x,a2) = // and &(//, cr2) = 1. Hence, the measure of distance

- 35 -
between [0] and [0i] is
¿[0i; 0] = inf ||/i - fi*||2 = ll/io - ^l2-
fc)l
The second infimum is zero since the model [0] is known to hold.
The measure of distance between [02] and [0] is
¿[02; 0] = inf \\fi - /z*||2 = inf \\X(3 - /z*||2
t>2 P
= I|X(X'x)->xv*-m*II2
(2.2.5)
= ||(/. - X(X'X)->X>'||2
= m*'(/„-x(x'x)-’x'K-
This is the squared length of the vector orthogonal to the projection of ¡i*
onto the range space of X. Notice that if ¿i* = X/3*, that is 02 holds, then
¿[e2;0] = O.
Finally, the distance between [02] and [02] is
¿[03;02] = inf ||/i - /i*||2 - inf ll/x - /¿*||2
= ^\in - ^ K - #**'(/„ - x(x'x)-’x>* (2.2.6)
= i¡’,\x(x'x)-1xl -
As another example, consider a random vector Y = (Yj,.. with
independent components following an exponential dispersion distribution
(Jprgenson, 1989). That is,
Yi ~ indep FJD(/ij,cr2), i = l,...,n,
where the density of Yi, with respect to some measure, has form
/v(y;7i,^2) = a(y,o-2)exp{^2(y7i - «(7i)>
(2.2.7)

- 36 -
where /z¿ = «'(7¿) and var(y¿) = cr2/í"(7¿). Let V(/i) = ®”/í"(7¿) and
# = (/ii,.. .,/xn,(j2)'. Since the components of y have different variances,
a natural measure of distance is
¿[0m2;©mJ = inf ||y(/i)-1/2(/i-^)||2-inf ||^(m)~1/2(^ - M*)||2- (2.2.8)
WM2
That is a(0) = /j, and b(0) = y(/¿)-1/2. Premultiplying the vector (/¿ - ¡j,*) by
Vr(/i)-1/2 has the effect of downplaying those differences (/q - ¿¿*) when the
corresponding variance is large.
To assess the goodness of fit of a model, relative to another, we can
estimate the distance 8 via some statistic based on the observed data. It
is interesting to note that when 8 = 0, i.e. both models hold, our data-
based estimate of this null distance will be some nonnegative (positive, if
the model is unsaturated) number, reflecting the amount of white noise or
random variability there is in Y. This is so because, if both models hold,
then the only reason that our estimate of distance would be nonzero would
be because Y has some random component. That is, the variability in Y that
is not explained by the model causes the data to fit the model imperfectly.
Let D be an estimate of 8. That is, -D[02; ©i] is a stochastic, data-based
estimate of how far apart models [©1] and [02] are. Potential candidates
for D are the weighted least squares, likelihood ratio, Wald, deviance, and
Lagrange multiplier statistics.
For example, consider the n-variate normal case and the four candidate
models [0], [0j], [02], and [03]. We will assume that both [0] and [02]
hold. In view of (2.2.5) a reasonable estimate of <5[02; 0] can be obtained by

-37-
replacing /¿* by Y, the estimate of under model [0], i.e.
D[02; 6] = Y'(I„ - X(X'X)-'X')Y = ¿(r¡ - i;)2.
1
Recall, that since ¿[02;0] is known to be zero, Z?[02;0] serves as our
‘estimate of error’.
Similarly, a reasonable estimate of ¿[03; 02] can be obtained by replacing
¡u,* in (2.2.6) by Y, the least restrictive estimate of /i*, i.e.
L>[03; 02] = Y'{X(X'X)-'X' - = ¿(Y, - Y)2.
Tl>
1
Now 03 C 02 and
d/[03] = n + l-(n-l) = 2
d/[02] = n + l-(n-p)=p+l.
The degrees of freedom associated with estimating the distance between
two models will be called the distance (or residual or goodness-of-fit) degrees
of freedom. The distance degrees of freedom for the two models [©Mi] and
[®m2] is defined to be the difference between the two model degrees of freedom,
i.e.
d/(¿[0M2; 0mJ) =d/[0MJ -d/[0M,].
The number of distance degrees of freedom measures the dimensional distance
between the two models, i.e. the difference in dimensions. It measures the
difference in the amount of freedom one has for estimating 9 for the two
models. It seems intuitive that if the degrees of freedom is large, that is the
dimensional difference between the two models great, the significance of the
distance statistic may be difficult to ascertain. This follows since we expect
the fit to be quite different for the two very different models, even when both

- 38 -
models hold. This is a reflection of both white noise and possibly lack of fit.
Therefore, the distance statistic will tend to be large, even when both models
hold. But for many statistics, a large mean implies a large variance, thereby
making significant findings more difficult. It is for this reason that we say
it is better to concentrate our efforts on relatively few degrees of freedom
to detect lack of fit. That is, one should use the smallest alternative space
possible when testing a null hypothesis.
A more technical argument holds when the test statistic (distance
statistic) is a Chi-square or an F. Das Gupta and Perlman (1974) showed
that for a fixed noncentrality parameter, i.e. fixed distance between models,
the power of the F-test or the Chi-square test increases as the distance degrees
of freedom decreases.
Example 1: Continuing with the n-variate normal example, we see that
¿mee,]) = df[@2] -d/[0s] = (p+i)-2=p-i.
Thus, 03 is of p - 1 less dimensions than 02. Now, if we knew a2 the white
noise variance, we could test H0 : 6* e @3, vs. Hi : 6* e ©2 - ©3, using the
statistic
¿>[83; 62] = SSjReg)
which has a X2(p-1) null distribution. However, a2 is not generally known and
we must estimate it. One way of estimating a2 is by estimating the distance
between [0] and [©2], two models that are known to hold, and dividing by
the distance degrees of freedom. Since the distance degrees of freedom is
df[Q] - df[Q2] = n-f 1 - (p +1) = n-p, we have that the estimate of the white
noise variance is D[02; ©]/(u-p) = SS(Error)/(n - p).

- 39 -
Notice that in the above example the estimate of the parameter a2
was simply the estimated distance between two models that were known to
hold divided by their dimensional distance. Quite generally, when the data
have an exponential dispersion distribution (2.2.7) with common dispersion
parameter a2, the estimated distance between two models that are known to
hold, divided by their dimensional distance gives us an estimate of a2. This is
true when the estimated distance is taken to be the LR, Wald, Deviance, LM,
or the weighted least squares statistics. These statistics are natural estimators
of the weighted distance given in (2.2.8) for the exponential dispersion models.
Now, let us assume that 0j and ©2 are each subsets of an (s -
l)-dimensional simplex. For example, with count data, conditional on the
total n, the distribution is often multinomial with index n and parameter
(alternatively, probability distribution vector) 0*. Read and Cressie (1988)
extensively study a family of distance measures called the power-divergence
family. The power divergences have form
Ix^=w+T)p;l{%Y-1h~ OO,
where Io and I-1 are defined to be the continuous limiting value as A 0 and
A —> -1. It is assumed that 9* and 8 fall on an (s - 1)-dimensional simplex.
As usual, let 0* represent the true unknown parameter. We define the family
of distance measures between [©j] and [©2] (©2 C0j) to be proportional to
6[©2;©i] = 2n{infJA(0*,0) -inf/A(0*,0)}.
©2 ©1
By properties of Ix{9*,0) (Read and Cressie, 1988, pp. 110-113), it follows
that 8 > 0, with equality if and only if both models hold.

- 40 -
To estimate ¿[©2; ©i] based on the data, we note that our least restrictive
guess of 9* is Y/n, the vector of sample proportions. Intuitively, a good
estimate of the quantity ¿[©2; ©1] would be
D[02; ©1] = 2n{ inf IA(Y/n, 9) - inf 7A(Y/n, 0)}
©2 ©i
= 1 yy.lY Yi Y-i] - yy-IÍ—í-Y- 1
A(A + l)^r*LU(A)J XJ A(A + l)tr*lU(A)j
where 0jA^ and 9^ are the ‘minimum divergence’ estimators obtained by
minimizing Ix(Y/n,9) with respect to 9 over 04 and 02 respectively. Read
and Cressie (1988) point out that Z)[02;0i] is equal to the likelihood ratio
statistic when A = 0. Also, if we assume that [O^ holds so that the second
infimum is zero, we have that, for A = 1,
(V. - n«,(1))2
0[e3;0.] = x;
TV
9^
which is asymptotically equivalent to
n[Q . 0 ] _ V'' O^i — n9\ ^)2
where 0(°) is the maximum likelihood estimator of 9* over the space 02. This
is the Pearson chi-square statistic. Other asymptotically equivalent distance
estimates are the Wald statistic and the Lagrangian multiplier statistic. We
now illustrate these results via examples.
Example 2: Suppose that Y = (Yn, Y12, Y2J, Y22)' is a multinomial vector.
That is,
(Yn,Y12,Y21,Y22)' ~ MuZ¿(n,(7rn,7r12,7r21,7r32)'), with = 1.
* j
Thus, the model that is known to contain the true parameter vector 7r* is [0]
where
© = (7T :7tT4 = 1,7Tij e (0,1), i,j = 1,2}.

-41 -
Notice that 0 is really a 3-dimensional subset (simplex) of (0, l)4 so that
d/[0]=4-l = 3.
We wish to test the independence hypotheses
Í H0 : 7rn7r22 = 7r127r2i, vs.
I H\ : 7T117T22 ^ 7T127T21
Writing the model of interest [0O] as
00 = {vr e 0 : 7T117T22 - 7T127T21 = 0}
= (tt : 7t'14 = l,7rn7r22 - ^12^21 = 0},
we can state the independence hypotheses as
Í H0 : 7T G 0O, vs.
1 #1 : 7T G 0 - 00-
Now, the model degrees of freedom can be found by subtracting the number
of constraints implied by [0O] from the total number of parameters, which
is 4. Hence, df[Q0] =4-2 = 2. Thus, the distance degrees of freedom or
measure of dimensional distance, is d/(<5[0o; 0]) = 3-2 = 1.
Two distance (goodness-of-fit) statistics commonly used are the Pearson
chi-square X2 (A = l) and the likelihood ratio statistic G2 (A = 0). The forms
of these two statistics are
D[e„;e] = *’ = ££
* Í
(yij - nTTijto)2
and
£[0o; e] = gj = 2 £ J»g (Jh.
i j n7r *J,0
where 7ris the ML estimate of 7r¿j assuming that model [0O] holds.
Under the null hypothesis, i.e. if independence truly holds, then the
asymptotic distribution of both distance statistics, X2 and G2, is X2(l)-

- 42 -
Example 3: Continuing with example 2, consider the model [®mh] where
®mh = {tt : tt'14 = 1, 7T1+ - 7r+1 = 0}.
This model implies that there is marginal homogeneity, i.e. The marginal
distributions for both factors are the same.
We would like to test the hypotheses
H0 : 7T g QMh, vs.
Hi : 7T G 0 - ®MH-
The model degrees of freedom is df[®MH\ =4-2 = 2, and so the distance
degrees of freedom is df(6[QMH] 0]) = 3-2 = 1. Once again, to illustrate
what model degrees of freedom means, we observe that if [®mh] holds and
we specify two of the four probabilities, the remaining two are completely
determined. Thus, we are free to estimate two of the probabilities based on
the data. The other two are determined.
Two frequently used estimates of the model distance, or model goodness
of fit are the likelihood ratio statistic G2 and the McNemar statistic M2. For
2x2 tables, the McNemar statistic and the Lagrange Multiplier statistic are
equivalent since both are score statistics (Agresti, 1990; Aitchison & Silvey,
1958). The statistics take the following forms
and
•D[eMiI;e] = G2 = 2£y>,iog(
* j
Vij \
nrriji 0}'
D[QMH-e] = Mi =
(yi2 - yii)2
yn + V2i '
where the 7r¿J)0 in the first expression is the ML estimate of 7r¿;- under the
model [0MJi].

- 43 -
Under the null, i.e. when the marginal distributions are homogeneous,
both of these statistics have asymptotic %2(l) distributions.
It is important to note that, had the constraint 7r2+-7r+2 = 0 been added,
the model would remain consistent but would be ill defined. For 2x2 tables,
this additional constraint is exactly the same as the constraint 7r1+ - 7T+1 = 0.
2.3 Multivariate Polytomous Response Model Fitting
In this section, we describe ML model fitting for an integer valued
random vector Y that is assumed to be distributed product-multinomially.
We also investigate the asymptotic behavior of the ML estimators within the
framework of constraint models. The models we will consider have form
Qx = {£ e 0 : Clog(Aeí) = X/3, Lé = 0}
or equivalently, for appropriately chosen U,
Qx = 0k = {(e@: U'C]og(Aé) = 0,Lé = 0},
where e¿ is the s x 1 mean vector of Y, a product-multinomial random vector
and the model parameter space 0 is of dimension s - q, where q is the number
of identifiability constraints. We use the parameter £ rather than fx — é
for several reasons. One reason will become evident when we explore the
asymptotic behavior of the ML estimator of £. It turns out that the random
variable ¡x - ¡u,o is not bounded in probability, whereas £ - £o is. In fact, the
random variable £ - £0 converges in probability to 0. Another reason for using
£ rather than ¡x is that the procedure for deriving the maximum likelihood
estimate of £ is less sensitive to small (or zero) counts. The range of possible
£ values is the whole real line, while the range of possible /x values is restricted

- 44 -
to the positive half of the real line. By using £ the problem of intermediate
out of range values (e.g. negative cell mean estimates) is avoided.
As stated above, we initially assume that the vector of cell counts Y
has a product-multinomial distribution. This is not overly restrictive since it
will be shown that inferences based on maximum (multinomial) likelihood
estimates are often the same as inferences based on maximum (Poisson)
likelihood estimates. We will present some results in section 2.4 that allow
us to determine when these inferences are indeed the same.
We also consider an alternative method for computing the maximum
likelihood estimators and their asymptotic covariances. The method of
Lagrange undetermined multipliers is well suited for maximum likelihood
fitting of the models we will be considering. This is so because we will specify
the models in terms of constraint equations and the fitting problem will be
one of maximizing a function, namely the log likelihood, subject to some
constraints, namely that £ 6 0/,.
2.3.1 A General Multinomial Response Model
In this section we specify a class of models that is directly applicable
to Chapter 3 of this dissertation. Specifically, the models will be specified in
such a way so as to include the class of simultaneous models for the joint and
marginal distributions considered in Chapter 3.

-45 -
Let the random vector Y = vec(Y’1,..., Yk) denote a product multinomial
random vector, i.e.
Yi = {Yu, ■ ■ ■ ,Yír)' ~ ind Mult(n¿,7r¿), i = K> 1,
where the R x 1 vector of cell probabilities satisfy k^Ir = 1, i = 1,..., K.
Consider the 1:1 reparameterization from {7r¿} to {£,}, where =
log(/¿¿) = log(n¿7Tj) is an R x 1 vector of log means. Under this parame¬
terization,
Yi ~ ind Mult(n¿, —), e^l# = n¿, i = l,...,K,
Tli
or
Yi ~ ind Mult(n¿, —), i = l,...,K, e^eflR) = n', (2.3.1)
Tli
where n' = (n1?... , n#) is the 1 x K vector of multinomial indices.
The kernel of the log likelihood for Y, written as a function of £, is
f(M)(£;») = „'£, e«’(©f 1*) = n' (2.3.2)
We now posit a model for £, the vector of log means. Let s = RK be the
total number of cell means. Our objectives are to test the model goodness
of fit and to estimate the s x 1 model parameter vector £ as well as any
freedom parameters of interest. It will be assumed that the model [©x] can
be specified as
©x — {£ £ Rs • Ci log A\^ =■ Xi(3i, C2 log — X2P2, Le^ = 0,
^'(©f 1r) = n'}>
(2.3.3)

- 46 -
where
Ci = ©fCtf, Cij = Cü, is qi xm¡ ¿ = 1,2
A¿ — A-ij, Aij = An, is m¡ x R, i = 1,2
L = = L\ is d x R
f = vec(6,---,^), and & is Jí x 1
X¿ is Kqi x pi of full rank p¿, ¿ = 1,2
n is the if x 1 vector of multinomial indices
s = RK, the total number of cells
Let us say that a model that can be specified as in (2.3.3) satisfies
assumption (Al). That is,
(Al) The multinomial response model can be specified as in (2.3.3).
Notice that the K matrices of Ct- are all identical, likewise with the
matrices comprising A¿ and L. This requires that the model does not change
across the K populations (K multinomials). Also, the two sets of freedom
equations in (2.3.3) will allow us to use two different types of models for
the expected cell means. This provides us with enough generality to fit
many interesting models. For example, we may wish to simultaneously fit
a linear-by-linear association loglinear model for the joint distribution and a
cumulative logit model for the marginal distributions.
We can conveniently rewrite (2.3.3) as
Qx = {ZeR’: Clog(Aeí) = X(3,L¿ = 0,e*'(©f 1*) - n'}, (2.3.4)
where A' = [A\, A'2\, C = C-y® C2, X = X\ ® X2, and ¡3 = vec(/31?/?2).
Notice that the model [0x] is specified in terms of both freedom
equations and constraint equations. We will rewrite [©x] as a constraint

-47-
model keeping in the back of our minds that the freedom parameters may be
of interest also.
Let U be a K(q\ + <72) x u matrix of full column rank u such that
U'X = 0. Here u is the dimension of the null space of X', Ai(X'), i.e.
u = K(qi + q2) - (pi +^2)- Since U can be chosen to be of full column rank, it
follows that the columns of U form a basis for the null space of X'. Thus, the
range space of U equals the null space of X', i.e. M{U) = X’{X'). Multiplying
the right and left hand side of the freedom equation Clog(Aeí) = X/3 by U1,
we can rewrite (2.3.4) as
eh = {£ e R° : C7'Clog(Aei) = 0,Le* = 0,e*'(®f 1R) -n' = 0}. (2.3.5)
Thus, 0x = ©h and the models [©x] and [©;,] are one and the same.
At this point, we will assume that the constraints implied by the model
[©/,] are nonredundant so that the model is well defined. More specifically, let
h'(£) = [(J7'Clog(Aei))',e^If'] be the 1 x (u + l) (l = Kd) vector of constraint
functions. We will assume that the u + / + K constraints implied by fi(£) = 0
and = n' are nonredundant. Notice that the constraints in fi(£) = 0
do not include the identifiability constraints. We treat the identifiability
constraints separately for reasons that will become apparent when we actually
fit the models.
As stated previously, one of our primary objectives is to estimate the
model parameters £ and the freedom parameters (3 under the assumption
that [©x] (and [©/,]) holds. We will use the maximum likelihood estimates,
which can be found by maximizing the log likelihood of Y subject to the
constraint that [0h] holds.

- 48 -
The (kernel of the) log likelihood under the product multinomial
assumption is shown in (2.3.2). It is
í(M)(í; y) = y'(-
Thus, we are to maximize the function y) = subject to £ e 0/,.
2.3.2 Maximum Likelihood Estimation
In this section we will discuss two procedurally different approaches
to maximizing the log likelihood subject to £ e Qh. The first
approach, which is the more commonly used approach, requires that the
model be specified entirely in terms of freedom equations. Often times,
when there are no identifiability constraints, the model can be completely
specified as a freedom model. Models amenable to this approach include the
Poisson loglinear model and the Normal linear model. The second approach,
Lagrange’s method of undetermined multipliers, can be directly applied when
the model is specified completely in terms of constraint equations. Since the
product multinomial model includes identifiability constraints, it can more
easily be specified in terms of constraint equations. For this reason this
second method is the preferred choice. In the following sections, we discuss
some additional features of these two methods.
Freedom Parameter Approach. One approach often used in simple situa¬
tions, namely those situations when the model can be specified completely
in terms of freedom equations, is to write the parameter £ as a function
of the freedom parameter (3 and maximize &M\£((3)\y) with respect to (3.
The vector £(/?) will be in the model space, since the model was specified

- 49 -
completely in terms of (3. For example, if the model could be specified as
0Jf = « € R-: loge< = X/3),
then £(/3) = Xf3. Notice that the multinomial model, which includes the K
constraints e^®^].#) = n', is not directly amenable to this approach. In fact,
we would have to reparameterize to a smaller set £* of s-K model parameters
that account for the K constraints. This reparameterization results in an
asymmetric treatment of the £ and for that reason is deemed undesirable.
On the other hand, the Poisson model considered below, will often lend itself
to this maximization approach, since the K constraints e^®^!.#) = n' are
not included.
Computationally, the method of maximizing the log likelihood with
respect to the freedom parameters is usually simple. Assuming the log
likelihood is concave and differentiable in /3, we need only solve for the root
of the ‘score equations’, viz.
s(/?;y)
SM-o
<9/3
Many of the asymptotic properties of the maximum likelihood estimator
/3 for /3 are derived by formally expanding the score vector s(/3;y) about the
true value /3 = ¡3* in a linear Taylor expansion. That is,
s(0- y) = s(/S*; y) + ds(^y)(H -/?*) + 0(11/3 - /3'll2)
In particular, in many situations,
0 = S0;Y) = S0-;Y) + 9s{Q,Y) 0 - F) + Of( 1),
(2.3.6)

- 50 -
so that (3 - ¡3* has the same asymptotic distribution as
)~V;n
Subsequently, we will derive the asymptotic distribution of $-(3* in a different
way. This alternative derivation of the asymptotic distribution of the freedom
parameter estimate will shed new light on the relationship between the
asymptotic behavior of the estimates under the two sampling assumptions—
product Poisson and product multinomial.
Expression (2.3.6) also gives some indication of how one might numer¬
ically solve for /3, the root of the score equation. A Newton-Raphson type
algorithm is often used. This root finding algorithm involves the inversion
of the derivative matrix ds((3;y)/d(3', which is usually of small dimension
since the model is usually specified in terms of a small number of freedom
parameters. In fact, the dimension of the derivative matrix will not be larger
than s x s, which occurs when the model is saturated.
Constraint Equations Approach. In many situations, it may be difficult to
specify a model in terms of only freedom parameters or perhaps it is possible
but the researcher would like to treat the model parameters symmetrically,
which would necessitate an additional constraint equation. It also could be
that the function ClogAe^ is not a 1:1 function of £ so that for given ¡3, we
can not solve for £ explicitly. In any of these cases, we may not be able to
use the aforementioned maximization approach.
In this section, we consider an alternative method for finding that £
that maximizes the function ^M^(£;y) subject to £ e 0j,. The method we
will use is the Lagrange’s method of undetermined multipliers. Aitchison and

- 51 -
Silvey (1958, 1960) and Silvey (1959) provide much of the essential underlying
theory related to this approach. Three positive features of this method
include (i) estimation of both £ and ¡3 is possible, (ii) the method provides
us with another enlightening way of deriving the asymptotic distribution
of the freedom parameter estimators, and (iii) the method works quite
generally. A negative feature of this approach is the computational difficulty.
Computationally, the method becomes burdensome as s, the number of log
mean parameters, and u + l + K, the number of constraints implied by the
model, become large. In fact, the algorithm involves the inversion of an
(s + u + l) x (s-\-u + l) matrix. One positive note, is that this potentially very
large matrix does have a simple form and one can invoke some simple matrix
algebra results to reduce the inversion problem to one of inverting matrices
of dimensions (u + 1) x (u + l) and s x s.
To best illustrate the difference in computational difficulty of the two
methods, we consider the following normal linear model example. Let
Yi ~ ind N(hí = (30 + PiXi,#2), i = 1,2,..., 100, a2 known.
The log likelihood can easily be written as a function of ¡3 = (/30,/3i)'.
Maximizing this likelihood with respect to (3 involves working with a 2 x 2
matrix. On the other hand, we could equivalently specify the linear model in
terms of the 98 constraints,
A*»+i ~ Hi = Vi+2 ~ IH+i i = i 2 .,98,
Xi+1 - Xi Xi+2 - Xi+1
and use Lagrange’s method. In this case, we would need to invert a matrix
which has dimension (s + u + l) x (s + u + l) = 198 x 198.

- 52 -
Even when we use the matrix algebra results that simplify the problem
of working with the 198 x 198 matrix, we still are left with a formidable task.
It seems that when s is large and the model is parsimonious, i.e. u + l + K,
the number of constraints is large, the undetermined multiplier method may
not be the method of choice. However, in time, as computer efficiency gains
are realized, we predict that the scope of candidate models to be fit using
this method will increase tremendously. In fact, at present, many categorical
models can easily be fit using Lagrange’s method. We discuss in more detail
how we can use the method of undetermined multipliers to fit models like
[0k] of (2.3.5).
We are to maximize the function y) = y'£ subject to the constraint
£ 6 0/,, where
0/, = {£ E Rs : C/'Clog^e^) = 0, Let = 0,e^(®^l^) - n' = 0}
= {ZeRs:h{t) = 0,et\(B?lR) = n'},
and h'(£) = [log(et'A')C'U, c?L%
Consider the Lagrangian objective function
F( 7) = f(M)(i;y) + (ef'(®f 1*) - n‘)T +
where 7 = vec(£,r, A). The K x 1 vector r and the (ii + /) x 1 vector A are
called either ‘Lagrange multipliers’ or ‘undetermined multipliers’.
Provided a maximum £ exists and that the Jacobian of [e^sfT#) -
n',/i'(£)] is of full row rank u + 1 + K for all £ 6 0/,, we can solve for the
maximum by solving the system of equations
dF( 7^))
dy
(y + L»(eí(M))( ®f l*)íW + H(£M)AM >
( ©f - «
V M^(Ai))
= 0
(2.3.7)

- 53 -
where the matrix H(£) = dh'(£)/d£. The Jacobian condition basically
requires the constraints to be nonredundant, thereby making [0/,] a well-
defined model.
From this point on, for notational convenience, the indices for the direct
sum ® will be omitted unless they are different from 1 and K.
We now require the matrices of models [0^] and [0/,] to satisfy some
additional conditions. Let us assume that
(A2) Either C¿ = Iq.K or C¿( © lm¿) = 0, i = 1,2
and
(A3) If Ci = Iq.K then M(X,) d M(elm.)
The assumptions require to be either a contrast matrix (rows sum
to zero), a zero matrix, or the identity matrix. If C¿ is the identity matrix,
it will be required that there exists a set of columns in that spans a
space containing the range space of ®fTmi. For most models of interest
these conditions are met. For example, any of the logit type models, such as
cumulative or multiple logit models, can be specified with C being a contrast
matrix. For loglinear models, the condition (A3) is met whenever the model
includes a parameter for each of the K multinomials.
The following lemma will be useful in showing that the maximum
likelihood estimates of £ and ¡3 are equivalent under both sampling schemes-—
product-Poisson and product-multinomial. The lemma will also enable us to
reduce the number of equations in (2.3.7) that must be simultaneously solved
when computing the maximum (multinomial) likelihood estimators.

-54-
Lemma 2.3.1. If the matrices of models [©x] and [©/i] satisfy (Al), (A2),
and (AS), then provided the model holds
(® i= (©i Proof. Using matrix derivatives (MacRae, 1974; Magnus and Neudecker,
1988), it follows that
H(£) = [D(é)A!D-l(A¿)C'U, D(e^)L'}
Thus,
(© rR)H(0 =
( ( © e« )[A1, A'2)D-' ( ^ ) (C? © C2)U, @e« L[
[(©eíi^íD-'í^eíJC! ( |(»iyci, (®1W)CSF. 0
= 0
The third equality follows since the model holding implies that ®e¿íL¿ = 0.
The sixth equality can be seen via the following argument.
If both C¿’s are contrast matrices, or zero matrices, then (A2) implies
that the matrix [(©lm,)^, (ffilmj)^] 1S the zero matrix. On the other hand,
if both C\ and C2 are identity matrices, then since the columns of U span
the null space of X', which, by (A3), implies that the columns of U span a
set contained in the null space of
fffilmxY
VffllmJ ’

- 55 -
we have that [( © l'mi), ( © l'm2)]U = 0. Any other combination of Cj and C2
can also be seen to result in the matrix equaling zero. m
The following theorem gives conditions under which we can find the ML
estimators of £ by solving a reduced set of equations. The smaller system of
equations no longer includes the identifiability constraint equations.
Theorem 2.3.1 Let vec(£(M), r(M\ \(M^) be the solution to (2.3.7).
Assuming that (Al), (A2), and (A3) hold, the sub-vector vec(£(M\ A(M1)
is the solution to the reduced set of s + u + l equations
+H(eM)) A(M)
hC^M))
= 0
(2.3.8)
Proof: Premultiplying the first set of equations in (2.3.7) by ©1'^, we arrive
at
(e l'*)y + ( e e 1*R + ( ® l's)tf(í<">)Á = 0 (2.3.9)
Now, (© l'ñ)y = n and (© l'fl)Z)(eí ~(M)1
be that ( © e^ )( © 1#) = D(n), the diagonal matrix with the multinomial
indices on the diagonal. Further, by Lemma 2.3.1,
( © l'iZ)Lf(£(M)) = 0. Therefore, (2.3.9) can be rewritten as
n + D(n) fW = 0,
which implies that f(M) = -Ik- Now, since the identifiability constraints have
been explicitly accounted for when solving for r^M\ we can replace of
(2.3.7) by -1^ and omit the identifiability constraints. Thus, vec(£(M), A(M1)

- 56 -
is the solution to the reduced set of equations
/'!/-eí<",+tf(í<">)Á<">'\ =0
V Ml(M)) )
This is what we set out to show. g
Before detailing the iterative scheme used for solving (2.3.8), we will
explore the asymptotic behavior of the estimator = vec(£(M\
within the framework of constraint models.
2.3.3 Asymptotic Distribution of Product-Multinomial ML Estimators
In what follows, we will assume that K, the number of identifiability
constraints, is some fixed integer, K > 1. We also will assume that the
asymptotics hold as n* = min{n¿} approaches infinity and that n* ~ n¿, i =
1 That is, we assume that the asymptotic approximations hold as
each of the multinomial indices get large at the same rate.
The derivation of the asymptotic distribution of will follow closely
that of Aitchison and Silvey (1958). Briefly, Aitchison and Silvey show that
if the score vector is op(n) and the constraints are such that the derivative
matrices #(£) and dif'(£)/<9£ have elements that are bounded functions then,
provided certain mild regularity conditions hold, the maximum likelihood
estimator £ is an n-1/2-consistent estimator of £0 and A is an n1/2-consistent
estimator of 0. They show that the joint distribution of (n*/2(£ - £o)>™-1^2^)
is multivariate normal with zero mean and covariance matrix
( B~l - B-'HiH'B-'Hy'H'B-1 0
V 0 (H'B-'H)-1
where B is the information matrix and H is the derivative of the constraint
function.

-57-
In our application, however, there are some minor changes. With the pa¬
rameterization we use, the information matrix is zero since the (multinomial)
log likelihood (2.3.2) is linear in the parameter £. This happens because the
identifiability constraints e^( ©^ 1^) = n' are ignored, to preserve symmetry,
when differentiating. Also, in our parameterization, the constraints are in
terms of e^, the components of which are e&¿ = n¿7r¿¿. Thus, the constraints
and the corresponding derivative matrices may not be bounded. For example,
a typical constraint is of the form Let — o. It follows that the components
of Let and the derivatives are increasing without bound as the multinomial
indices are allowed to increase without bound.
Fortunately, we can still use the results of Aitchison and Silvey (1958)
by replacing the matrix H and the vector A/n of Aitchison and Silvey by
our H/n* and A, where n* = min{n¿}. The zero information problem can be
solved by identifying the vector Y - et as the ‘score vector’. It is pointed out
that, in this case, the asymptotic variance of ®D_1/2(n¿l^) times the score
vector is not equal to the negative derivative matrix D(7T0) but instead is
equal to D(tt0) - ©7ro¿7ró¿. This happens because the components of Y are not
independent; Y is product multinomial. Using this reparameterization, all of
the necessary assumptions required by Aitchison and Silvey (1958) hold, i.e.
assumptions X and H of Aitchison and Silvey (1958) hold.
As previously mentioned, Aitchison and Silvey show that A is an
n1/2—consistent estimator of 0. With our paramterization, having replaced
A/n by A, it follows that A(M) will be n^2-consistent. We now derive the
asymptotic distribution of

- 58 -
Define the stochastic function g by
MO
The maximum likelihood estimator is the solution to g[6\ Y) = 0.
Under our parameterization, using the results of Aitchison and Silvey
(1958), we have that each of the following hold
ef'"’ - e®> = - f„) + 0P(1),
H(| and
Thus,
M£(M)) = Mí.) + W'(Í»)(Í(M) - ío) + Op(l)
= tf'(£o)(Í(M)-ío) + Op(l),
ff(f 0 = s(¿(M);y)=('r-eí("l+íf(í(«))X(«)'j
V MÍ can be rewritten as
0=(Y~eí°- ■D(ef0)(f + Op(l)\
V íT'(ío)(f = ('r-eí»\ / D(eío) -H(í.)Uí(M)-íoUop(l)
l 0 ) „) 0 H Á(«) )+L,pW-
Therefore, it follows that
oD-'i’ím i*)
r-ef"
0
since n* ~ n,-, ¿ = 1,... ,K and 7r0 = ( ® -D-1(n¿lfl))eío.
(2.3.10)

- 59 -
Now, the random variable ®D_1/2(niliZ)(y-e¿°) is a vector of normalized
sample proportions so that
©JD-1/2(nil*)(y-eio)^
has an asymptotic normal distribution with zero mean and covariance matrix
^D(tTo) - ©XoiTT^ 0^
Therefore, by an extension of a theorem of Cramer (1949) and by equation
(2.3.10), it follows that nlJ2{6(M) - 9*) = n;¡/2vec(£(M) - £o>A(M)) has an
asymptotic normal distribution with mean zero and covariance
( D(ir„) (D(wh) - ®iroiic'e
{-*** o ) l 0
* í)(S T
This covariance matrix is shown in the appendix to have the simple form
Mi 0
0 M2
where
M, = D~\*0) - D-¡(k0)H(H'D-'(*0)H)-¡H'D-¡(vo) - ffif lal'a
and
Finally, using the fact that n* ~ n¿, i = 1,..., K, we can discriminantly
replace n* by the appropriate to arrive at a simple, asymptotically
equivalent, expression for the asymptotic covariance of = vec(£(M), A(M1).

- 60 -
It is
( D-' - D-'HiH'D-'Hy'H'D-1 - 0
V 0 ' (H'D-'H)-J
where Z? — 0) = D(e¿0) and H = if(£0)-
2.3.4 Lagrange’s Method—The Algorithm
In this section, we give details of how one can actually fit the models
of (2.3.4) or equivalently (2.3.5). We show how Lagrange’s undetermined
multipliers method can be used in conjunction with a modified Newton-
Raphson iterative scheme to compute the ML estimators and their asymptotic
covariances. We will assume that the model assumptions (Al), (A2), and (A3)
hold. This section includes an outline of the algorithm used in the FORTRAN
program ‘mle.restraint’.
Recall that our objective is to find that e Qx, where
Qx = iteR’: CTog(Aeí) = X/3, = 0, (®l'*)e« = n},
that maximizes the multinomial log likelihood
(2.3.12)
¿(M)(Í; y) = y'l
Since the assumptions (Al), (A2), and (A3) hold, we see by Theorem
2.3.1 that our problem is reduced to one of solving the system of equations
(2.3.8), i.e. to find the ML estimator = vec(^(Afl, we must
simultaneously solve the system of s + u + 1 equations

- 61 -
where the (u + I) x 1 vector h and the s x (w + /) matrix H are defined as
follows.
Mf)=([Wlog(¿eí))
and
m =
mt)
dZ â– 
It will be shown in section (2.4) that g(9) is actually the derivative
of the Lagrangian objective function under the product-Poisson sampling
assumption.
The iterative scheme used in the FORTRAN program ‘mle.restraint’ is
a modified Newton-Raphson algorithm. The algorithm can be sketched as
follows.
(1) Find a starting value for 8.
(2) Replace 0M by 9^) = - G“1(0('%(0<*')) (2.3.13)
(3) If ||5I(^^I/+1^)|| > tol go to (2). Else stop.
The matrix G{8) used in step (2) is actually
G(9) =
m
0
and the inverse of G{9) is of the very simple form (see Aitchison and Silvey,
1958 or Rao, 1974)
G~\9)
(D-1 - D-'HiH'D-'Hy'H'D-1 -D^HiH'D^H)-1 \
-(H'D-'Hy'H'D-* -(H'D-'H)-1 ) '
(2.3.14)

- 62 -
where D = D(et). Since we use G(0) in place of the Hessian matrix, the
procedure is a modification to the Newton-Raphson method. Haber (1985a)
used the more complicated Hessian matrix.
Notice that the inversion of G, which may be performed at each iteration,
is not nearly as difficult as inverting a general matrix of dimension (s + u +
l) x (s + u + /). First of all, in view of (2.3.14), to obtain the inverse of the
partitioned matrix G, we need only invert the matrices D and H'D~1H, which
are of dimension s x s and (u + l) x (u + Z). Secondly, the inversion of D is
simple since D is a diagonal matrix with e¿ on the diagonal. Hence, the most
formidable task in the inversion process is the inversion of the symmetric
positive definite matrix H'D^H. There are many efficient ways to invert
large symmetric positive definite matrices.
Upon convergence of the algorithm (2.3.13), estimates of the asymptotic
covariances of and A(M) are readily calculable. Write G_1(0) of (2.3.14)
as
where
P = D~1 - D-'HiH'D-'Hy'H'D-1
Q = -D~'lH(H'D-1H)-1
R = -{H'D-1H)-1
By (2.3.12), the asymptotic covariance of 9(M) = vec(£(M), AW) can be
estimated by
0 )
0 -R)
Variance estimates for other continuous functions of 6^M\ such as
¿(M) = and p(M) _ (X'J5f)-1X'Glog(Aeí(M)), can be found by invoking

- 63 -
the delta method. For example,
var )var (¿(M))D(ei<">)
and
var(/3(M))=
(X'X)-1X'CD-1(Afi(M))A(vax(fiM))A'D-1(Afi(M))C'X(X'X)-1.
Evidently, Lagrange’s method of undetermined multipliers provides us
with a convenient procedure for maximum likelihood fitting of models in a
very general class of parametric models for multivariate polytomous data with
covariates possible. We now briefly outline the steps needed to perform the
iterations of (2.3.13).
Computing U. The first thing we must do is write the freedom model (2.3.4),
which can easily be input by the user, as a constraint model (2.3.5). Therefore,
we must compute a full column rank matrix U that satisfies U'X = 0. The
method we use to find U is attributed to Haber (1985b).
Using the notation of ‘mle.restraint’, let X be a full column rank matrix
of dimension q x r. Let u = q - r be the dimension of the null space of X'.
Further the matrices A and C of (2.3.4) will have dimensions mxs and q x m
respectively. The relationship between these dimension variables and those
used in sections 2.3.1 and 2.3.2 is as follows
q = K(q1+q2)
r=p! +p2
m = K(rrii + m2).
We use the variables g, r, and m for notational convenience.

-64-
Consider the matrix U* = Iq-X(X'X)_1X'. This qxq matrix is of rank
u = q-r and satisfies the property
XJ*'X = 0.
Let W denote a q x u matrix with random elements. Specifically,
Wij ~ Uniform(0,100), i=l,..., It follows that the matrix W is of full column rank with probability one and
hence that the qxu matrix U = U*W is of full column rank u with probability
one. But the matrix U satisfies
U'X = W'U*'X = W'O = 0.
Therefore, at least with probability one, we have found a full column rank
matrix U that satisfies the property U'X = 0. Using this U, we are able to
write freedom model (2.3.4) as a constraint model (2.3.5).
Computing h(£). We write the constraint model of (2.3.5) as
{£efl':A(f) = 0, e<'(@fl*)=n'}, (2.3.15)
where the constraint function h is defined as
m=(u'c ^ef)).
Computing g(9). Notice that since (Al), (A2), and (A3) hold, the
identifiability constraints present in the product multinomial model (2.3.4)
can be accounted for explicitly. It will follow by results of section 2.4, that
under either sampling scheme—product-Poisson or product-multinomial—

- 65 -
the maximum likelihood estimators for £ and A can be found by solving the
equation
s(0)=(y-ef+^(fM)=°, (2.3.16)
where the matrix H is the derivative of h' with respect to £.
Computing #(£). We will use matrix derivative results of MacRae (1974)
to find the matrix of derivatives of the constraint function h'(£).
H(0 = ^ = ^[log(e = [D(é)A!D-\A¿)C'U, D(é)L'].
The equality follows upon using the matrix version of the chain rule. Notice
that
^(log(e‘,A')CC/)=(^)!j(be = D(é)A! D-\A¿)C'U
and that
dtPV dtP dtP U
dfi
8^ det
= D(et)L'.
Computing G{9). The iterative scheme (2.3.13) used to solve the system
of equations (2.3.16) is actually a slight modification of the Newton-Raphson
algorithm. It is a modification because we do not use the derivative matrix
G* = dg(9)/dd to adjust at each iteration, as Haber (1985a) did, but rather a
simpler matrix G that is related to G* by G* = G + Op(n\'2). The derivative

- 66 -
matrix G* can be computed as follows.
_ QgW _ \SaW Mill
' 1 ~ a» ~[ sf ’ d\‘ .
= i-D(e() + ^p H(()\
V H’(() o )
=(-hdW TWT
The matrix
dH(QA _ dH{Q
d?
{I, ® A)
is of order Op(i7.y2) when it is evaluated at 0 = vec(£, X) since
mi)
d?
0P(n*)
and
A = Op(n“1/2).
It follows that the matrix G, which is much simpler to invert than G*, can
be used to adjust the estimate at each iteration.
Computing the inverse of G. Although the matrix G is of dimension
(s + u + V) x (s + u + Z), which may be very large in practice, its inverse
is relatively simple to calculate. The inverse of the partitioned matrix
is shown by Aitchison and Silvey (1958) to have form
( D-1 - D-'HiH'D-'Hy'H'D-1 \
\ -(H'D-'Hy'H'D-1 -{H'D-'H)-' )’
Therefore, only the matrices D and (H'D 1H), which are of dimensions
s x s and {u + l) x {u + /), need to be inverted. The inverse of D is easily

-67-
calculated since D is a diagonal matrix with e* on the diagonal. The inverse
of (H1 D~JH), a symmetric positive definite matrix, can be found quite easily,
even when u + /, the number of constraints, is large. It should be pointed
out that when s, the total number of cell means, is large, the number of
constraints u +1 may be large and on the same order as s. This will be the
case for parsimonious models—those models with many constraints relative
to number of model parameters.
One could choose to invert the matrix G a limited number of times to
mitigate the computational burden. In fact, in their 1958 and 1960 papers,
Aitchison and Silvey advocate an iterative method whereby the inverse of G
is computed only two times. Once at the initial iteration and again at the
final iteration, upon convergence. We feel, however, that in this special case
in which the matrix G has a particularly simple form, the inverse can be
computed at each iteration. Along with increased computing power, there
are many efficient algorithms for inverting large symmetric positive definite
matrices.
2.4 Comparison of Product-Multinomial and Product-Poisson Estimators
We begin this section by introducing notation for a product-Poisson
random vector.
The sxl random vector Y = vec(Yi,..., YK) is said to be product-Poisson
if
Yij ~ ind Poisson(e^), i = l,...,K, j = 1,...,R. (2.4.1)
Suppose that the s = RK log means {&_,•} satisfy the model [0^] where
&P = {£eR': Clog(Aeí) = X/3, Le* = 0}

- 68
or equivalently, for appropriately chosen U,
&P = Qp = {£eRs: U'C\og(Ae*) = 0, = 0}
(2.4.2)
This model implies all the same constraints on £ as the product-
multinomial model [0/,] of (2.3.5), with one exception—the identifiability
constraints, e^( © 1#) = n', are not included.
Denote the maximum likelihood estimators computed assuming (2.4.1)
and (2.4.2) by £(p) and /3(ph Similarly, denote the maximum likelihood
estimators computed assuming (2.3.1) and (2.3.5) by £(M) and
Recall that the three product-multinomial model assumptions are
(Al) The multinomial response model can be specified as in (2.3.3).
That is the model parameter space can be represented as
0jt = {£ e R* : Ci logAje* = X\f3\,C2 log A2e^ = X2f32,
Lé = 0, ei'(©fl^) = n'},
where
Ci = ®f Cy, Cij = Ci 1, is qi x rrii 7 — 1,2
Ai = ©f Aij, Aij = An, is rrii x R, 7 = 1,2
L = Lj, Lj = Li is d x R
£ zz vec(£i,...,£^), and £*. is A x 1
Xi is Kqi x Pi of full rank p¿, 7 = 1,2
n is the AT x 1 vector of multinomial indices
s = RK, the total number of cells.

- 69 -
(A2) Either = Iq.K or C¿( 0 lmj) = 0, ¿ = 1,2,
and
(A3) If Ci = Iq.K then M(Xt) D M(elmi).
The following theorem states that the maximum likelihood estimators for
£ and hence ¡3 are the same under the product-multinomial sampling scheme
of (2.3.1) and the product-Poisson sampling scheme of (2.4.1) provided that
the three assumptions (Al), (A2), and (A3) hold.
Theorem 2.4.1 If the model (2.3.4) satisfies assumptions (Al), (A2), and
(AS), then
¡(P) = £W and £(p) =
That is, the maximum likelihood estimators of (3 and £ are the same under
both sampling schemes—product-Poisson (2.4-1) and product-multinomial
(2.3.1).
Proof: Under the product Poisson assumption of (2.4.1) and (2.4.2), the
kernel of the log likelihood is
¿(p)(£;y) = y'£-e*'i,.
Therefore, letting 9 = vec(£,A), the corresponding Lagrangian objective
function is
W)=»’í-e«,l. + W(í)A
and so to find the maximum (Poisson) likelihood estimator 9^ = (£(p), A(pl)
we must solve the system of equations
dQ(9) = (y- ei(p) + if(£(p))A(p)
90 \ h@p))
The conclusion of the theorem now follows, since the equations (2.3.8) of
\ =0. (2.4.3)

- 70 -
Theorem 2.3.1 and (2.4.3) yield exactly the same solutions and
/?(p) = (X'X^X'Clog^Ae^) = (X'Xy'X'ClogiAe^) = ftM\
As a corollary to Theorem 2.4.1 we have
Corollary 2.4.1 Provided the assumptions of Theorem 2.f.l hold, the
estimated undetermined multipliers are invariant with respect to sampling
scheme, i.e.
A(m) = X(P)
Proof: The proof follows immediately upon noting that equations (2.3.8)
and (2.4.3) yield exactly the same solutions. _
A remark is in order. Basically, Theorem 2.4.1 enables us to conclude
that the sufficient and necessary condition of Birch (1963) holds. These
conditions are that the model be specified so that the Poisson ML estimators
necessarily satisfy the identifiability constraints that are required for the
multinomial model.
We now explore the asymptotic behavior of the (Poisson) ML estimator
0(p) = vec(£(p\ A(p)). For the product-Poisson assumptions (2.4.1) and
(2.4.2), we can obtain the asymptotic distribution of Q(p) by formally replacing
the n* = min{n¿} by ¿¿* = min{e&>} and using the same arguments as those
used to derive the asymptotic distribution of 0(M).
Jprgenson (1989) discusses limiting distributions for Poisson random
variables as the mean parameters, or equivalently /i*, go to infinity. In this

- 71 -
case,
al»';Y)=(r- + )
has an asymptotic normal distribution with mean zero and asymptotic
covariance
(^°)
Using arguments similar to those used in the multinomial case, it follows that
(Y-e*°
V 0
-H
0
^ ( £(p) - 6
/ l A(p)
We conclude, as in the product multinomial case, that - 90 has an
asymptotic normal distribution with mean zero and asymptotic covariance
(DM -Hy'fD^o) o) /%) -fry-1
\ -H' 0 ) \ 0 0) [ -H' 0 ) •
But, this can again be simplified as it was in the multinomial case. It can be
shown that the asymptotic covariance can be rewritten as
(D-* -D-'HiH'D-'Hy'H'D-1
V 0
0 ^
(.H'D-'H)-J)
(2.4.4)
where D = D(/j,0) = D(eio) and H = H(£0)-
Comparison of the Asymptotic Distributions. Provided assumptions (Al),
(A2), and (A3) hold, both 9— 0O and - 90 have asymptotic normal
distributions with zero means and respective covariances given in (2.4.4) and
(2.3.12). Therefore, we have the following interesting results.
Result 1. The asymptotic covariances of ¿Ap) and are related by
-(•t;)
var( iVÍ*) = var(#(i’*)
(2.4.5)

- 72 -
Result 2. The asymptotic distributions of A(p) and A^) are identical and
it follows that the Lagrange multiplier statistic which has form
LM = A'(var(A))-1A = A \H'D-'H)\
is invariant with respect to the sampling scheme.
Result 3.
~(P)~(P)'
var= var(/¿(p)) - ———
Tlx
(2.4.6)
Result 4.
var(/?(M)) = var(/?(p)) - A (2.4.7)
where
A = (X'X)~1X'C
Q^)C'X(X'X)-1.
VUi
and is nonnegative definite.
The notation var(-) used in these results denotes the asymptotic variance.
This is important since the finite sample variances may not even exist.
The proofs for Results 3 and 4 are straightforward. Basically, they
involve using the delta method and equation (2.4.5). The interested reader
will find an outline of the proofs in Appendix A.
In practice, it is of particular interest to evaluate the matrix A of equation
(2.4.7). Often, for convenience, the models are fit assuming the vector Y
is product Poisson and then inferences based on the maximum likelihood
estimates are made assuming that they are invariant with respect to the
sampling assumption. Birch (1963) and Palmgren (1981) derive rules for

- 73 -
when these inferences, based on the two different sampling assumptions, will
be equivalent. However, they assume that the model is of a simple loglinear
form. That is, the Poisson model is assumed to have form
0* = {£ G Rs : t = X(3}.
We will use the results of this section to derive more general rules for when
the two inferences will be equal. As a special case of these results, we will
arrive at the Birch and Palmgren results.
The following lemma will enable us to rewrite A of (2.4.7) in still a
simpler form.
Lemma 2.4.1 Let Z = [Z\,..., Zk\ be an r x K matrix of full rank K.
Suppose that X = [Xx,..., Xp\ is an r x p (r > p > K) matrix of full rank p
such that m(X) D m(Z), i.e. the range space of X contains the range space
of Z. Denote the T (K M(Z) by {X^,... ,X„T}. Without loss of generality, suppose that the set of
vectors {XVl,...,XVT} is a minimal spanning subset, i.e. the spanning set
of any r < T of these vectors does not contain the range space of Z. We
conclude that
3W e RTxK 3 (X'X)~lX'Z = JW,
where the p x T matrix J = [eVl,..., e„r] and eVi is the p x 1 vector
(0,..., 0,1,0,..., 0)' with the ‘1 ’ in the v\h position.

-74-
Proof: Let X* = [XVl,... , X„T]. Now, by assumption, M(X*) D M(Z).
Hence, there must exist a matrix W e RTxK 3 Z = X*W. Therefore,
(x'xy'x'z = (x'xy'x'XtW = (x,x)-1(x,x*)w’ = jw
where J = (X'X)_1(X'X*) is as stated in the conclusion of the lemma. g
Before stating the next important theorem, let us write A in another
way. Assuming that (Al) holds, A can be written as
a=(a“ A22) (2.4.8)
where
Ay = [XW'XM @ ^)( ® ^-WXjiXW'.
Now, if C¡ is a contrast matrix, by assumption (A2), we can write
(XIXJ-'XICJ © lp~) = 0 = JWwW,
vnk
(2.4.9)
where Jb) can arbitrarily be chosen to be equal to X[ and so Wb) = 0. On
the other hand, if C, = Iq.K then we have by (A3) that At(X¿) D Xí(®lm¡).
Therefore, we can invoke the result of Lemma 2.4.1 by setting Z =
Since M(Xi) 2 Al(ffilm¡) = M(Z), the conditions for the lemma are satisfied.
Let X;* = [X. (,),...,X. (,•)] be the x (K < Ti < Pi) submatrix of X¿
W1 tVT¡
that has columns that form a minimal spanning subset for M(Z) — A^©-^»-).
By Lemma 2.4.1,
3W® 6 RTixK 3 (X!X¿)_1X!( ffi Ip-) = JWW&.
Vnk
(2.4.10)
Here, Jb) = [e (i),..., e (i)], where the T¡¡ elementary vectors correspond to the
V1 UTi
columns (X. X. (i)} of X¿ that form a minimal spanning subset for the

- 75 -
range space of ®lm¡, i.e. the columns span a set that contains the range
space of ®lmi and any smaller set of columns will not span a set containing
the range space of ©lm¡ .
It follows that the matrices A,J of (2.4.8) can be written as
A*-7 = jWwWw'O) J'U)
where
[e (i)], if Ci=Iq.K
X[, otherwise
and
= if Ci =/« jt
\ 0, otherwise.
We now state a theorem of substantive importance.
Theorem 2.4.2 Suppose that assumptions (Al), (A2), and (AS) hold. For
r = 1,2, if Cr is the identity matrix then let {u[r\... ,u^} be the set of
indices that index those columns of Xr that form a minimal spanning subset
for Ai(ffilmr). Then it follows that the relationship between the asymptotic
variances of the two estimators /3(M) and ¡3(p) is
var(/?(M)) = var(/?(p)) - ^22 ) ,
where the pi x pj matrix AlJ is a zero matrix whenever at least one of or
Cj is a contrast or zero matrix. Otherwise, if both C¿ and Cj are identity
matrices then
(2.4.11)
(2.4.12)
(2.4.13)
^ki —
if (k,l) (f {v\
(*)
ut!}
X

-76-
Proof: Since (Al), (A2), and (A3) hold, we can rewrite A,JI as in (2.4.11).
Now, if either C¿ or Cj are contrast or zero matrices, it is obvious by (2.4.9)
that A*-7 will have zero components, as stated in the theorem, since at least
one of jyb) or Wwill be a zero matrix. On the other hand, if both C¿ and
Cj are identity matrices, then A1-7 can be rewritten as in (2.4.11) where
and the matrices Wb) and are elements of RT¡xK and RTixK. Hence,
V
where W'i = W^W'C) is some xTj matrix. Now, since {e„} are elementary
vectors, we have that if
(M) £ K
(i)
Ü)
’ UT,
},
then the component A^t = 0. Otherwise, if (k,l) is a member of this set, it
must be that A¿j is one of the elements of the matrix W,J. This completes
the proof. ^
The next two corollaries follow immediately from Theorem 2.4.1.
Corollary 2.4.2 If both C\ and are contrast matrices then
var(^(M^) = var(^(p)).

-77-
Proof: Since both Cj and C2 are contrast matrices it follows that W(*)
and WW are zero matrices. Therefore, the matrices A‘J of the theorem are
zero matrices. g
Corollary 2.4.3 Let C2 = 0,X2 = 0, and C\ = A\ = I3, so that the model
(2.3.4) becomes
e* = {( 6 R‘ ■■ £ = xp, e<'(els)=n'>,
i.e. a simple loglinear model with K subpopulations. Let {v\,..., Ut} be the set
of indices that index the columns of X that form a minimal spanning subset
for Then
var(/3(M)) = var(/5(p)) - A,
where the elements of A are such that
A-ki = 0, if (M) 0 M,...,^r}2
Proof: The proof is an immediate consequence of the theorem upon
identifying A11 of the theorem with A of the corollary. The other matrices
A12, A21, and A22 will be zero since C2 = 0. _
Corollary 2.4.3 is of practical importance and is essentially the result
shown by Palmgren (1981). In particular, if we parameterize the model in
such a way so that there is a parameter included for each of the K independent
multinomials (or K covariate levels), then the K columns of X corresponding
to these K ‘fixed by design’ parameters will form a basis (and hence a minimal
spanning subset) for At(®f lj?). Therefore, if (3i and (3j are not one of the

- 78 -
K parameters fixed by design, then cav(^M\^M^) = cov{j3\p\j3^). We
will illustrate the utility of the above results in the next chapter of this
dissertation.
The next section considers issues that may arise when computing the
model degrees of freedom. It also states some other miscellaneous results
with regard to the Lagrange multiplier statistic.
2.5 Miscellaneous Results
We begin this section by addressing practical issues that may arise during
nonstandard model fitting. Specifically, we will consider computing the model
and distance (or residual) degrees of freedom.
Computing model and distance degrees of freedom. Assuming the model
[0ft] of (2.3.5) is well defined, i.e. the u + l + K constraints are nonredundant,
we can compute the model degrees of freedom as in section 2.2. In that
section, we defined the model degrees of freedom as the number of model
parameters minus the number of independent constraints implied by the
model. Notice that in this application we have an additional l linear
constraints. The l constraints were not present in section 2.2. It follows
that the model degrees of freedom for [©/,] is
df[Qh] = s - (u + l + K) (2.5.1)
where s is the number of cell means, u is the dimension of the null space of X',
l is the number of linear constraints, and K is the number of identifiability
constraints.

- 79 -
To measure model goodness of fit, we can consider estimating some
hypothetical distance between model [0/,] and the saturated model (u = l = 0)
[0]. This distance, denoted <$[0/,;0] has degrees of freedom
d/(i[ek;e]) = d/[0]-d/[e4]
= (s-K)-(s-(u + l + K)) (2.5.2)
= u + 1.
Notice that, had we considered the product Poisson model (2.4.2), the
distance degrees of freedom would be
df(8[Q^; 0(p)]) = s - (s - (u + /)) = u +1,
which is identical to the product multinomial distance degrees of freedom of
(2.5.2).
We have assumed that the u + l + K constraints are nonredundant, i.e.
each constraint is not implied by the other constraints. This may not always
be the case. To illustrate, consider the model specification for example 3 of
section 2.2.2. The model [®mh] implies that the two marginal distributions
are equal. We stated at the end of that example that the additional constraint
7t2+ - 7t+2 = 0 was redundant. This can be seen since
7r2+ ~ 7r+2 — Tn - 7r12 = -("^l^- - tt+i) — 0
That is, the constraints of model [®mh] imply that 7r2+ - 7r+2 equals zero.
Had we blindly added this constraint, we may have incorrectly calculated
the model degrees of freedom as 1 and the distance degrees of freedom as 2.
Therefore, we must be very careful to have a set of nonredundant constraints
when computing degrees of freedom.

- 80 -
In practice, when models are more complicated, it may be difficult to as¬
certain whether or not the model constraints are nonredundant. Fortunately,
there are two very useful results that help in this regard.
The first result is that when the constraints are redundant, the matrix
evaluated at some point in 0/, is of less than full rank and is not
invertible. Therefore, in practice, if the algorithm (2.3.13) does not converge
due to G being singular, it may be due to redundant constraints, i.e. an ill-
defined model. The user should investigate and possibly respecify the model
should this occur. A caveat is that due to computational roundoff error, a
singularity may not occur even when the model is ill defined because the
iterate estimates, including the final estimate, may not strictly lie in Qh. The
next result may mitigate this problem.
A result that is useful in practice is that a necessary condition for the
constraints to be nonredundant or equivalently for the model to be well
defined, is that the Lagrange multiplier statistic be invariant to choice of
U, a matrix with columns spanning the null space of X1. Evidently, if the
user fits the model several times, each time using a different lU' matrix, and
the Lagrange multiplier statistic varies (more so than can be explained by
roundoff error), then it must be that the model is ill defined.
Formally, this necessary condition can be stated as
Theorem 2.5.1 Let U\ and U2 (U\ U2) be any two full column rank
matrices satisfying U-X = 0, ¿ = 1,2. Denote the Lagrange multiplier statistic
evaluated using Ui by LM{Ui). If the matrix
Hi = A’)CUi, é‘U)

- 81 -
is such that [Hi, e^] is of full column rank, i = 1,2, and hence the models well
defined, then
LM{UX) = LM(U2),
i.e. the value of the Lagrange multiplier statistic is invariant with respect to
choice of U.
Proof: Denote the model specified in terms of C/¿ by [0fc.], i = 1,2. By
the definition of U{ we know that the constraints implied by [0ftJ and [0ftj]
are equivalent. Hence, the solution £ to (2.3.8), or equivalently (2.4.3), under
either model is the same. Thus, in view of the first set of equations in (2.3.8),
any solution vec(£, A¿) under model [0/,,.] must satisfy
-(y-e¿)= HiCOk, ¿ = 1,2. (2.5.3)
Notice that since Ux ^ U2, we have that Hx{\) ^ H2(£) and by (2.5.3) Aj ^ A2.
Now, (2.5.3) implies that
ffi(É)Á, = H2{t) A2.
(2.5.4)
Also, since üf¿(£) is assumed to be of full column rank, the variance of Á¿,
var(Á¡) = (ffKéjIT'íeíWí))
-1
(2.5.5)
exists. Therefore, the Lagrange multiplier statistics LM{Ui), which have form
Á'[var(A¿)] 1Xi, i = l,2
(2.5.6)

- 82 -
exist. Finally, by (2.5.4)-(2.5.6), it follows that
LAf(CTi) = A^var^)]-1^
= X2(H'2(()D-'(¿)H,(0)M
= Á'2 [var(Á2)]Á2
= LM(Ut).
This completes the proof. g
The final result of this section states that the Lagrange multiplier
statistic is exactly the same as the Pearson chi-squared statistic whenever the
random vector Y is product-Poisson or product-multinomial and the model
satisfies assumptions (Al), (A2), and (A3).
Theorem 2.5,2 Assume that the product-multinomial model satisfies
assumptions (Al), (A2), and (A3). Let X2 denote the Pearson chi-squared
statistic, i.e.
X2 = (y- - fi)
where fi is the ML estimator under either of the sampling schemes—product-
multinomial or product-Poisson. It follows that the Lagrange multiplier
statistic LM is equivalent to X2. That is,
LM = X2.
Proof: By equations (2.5.3), (2.5.5), and (2.5.6) of the previous theorem’s
proof and the fact that e* = ¡x, we have that
LM = (y-¡i)'D-'Ui)(y- £) = *2
This is what we set out to show.

- 83 -
2.6 Discussion
In this chapter, we discussed in some detail issues related to parametric
modeling. In particular, we followed the lead of Aitchison and Silvey (1958,
1960) and Silvey (1959) and described two ways of specifying models—using
constraint equations and using freedom equations. In section 2.2, distance
measures for quantifying how far apart two models are, relative to how close
they are to holding, were discussed. In particular, the power-divergence
measures (Read and Cressie, 1988) were used when the parameter spaces were
subsets of an (s - l)-dimensional simplex. Estimates of these distances were
developed based on very intuitive notions. Also, a geometric interpretation
of model and residual (or distance) degrees of freedom was given.
In section 2.3, we described a general class of multivariate polytomous
(categorical) response data models. The class of models, which satisfy
assumptions (Al), (A2), and (A3), were shown to satisfy the necessary and
sufficient conditions of Birch (1963) so that the models could be fitted using
either the product-Poisson or product-multinomial sampling assumption.
An ML fitting method was developed, using results of Aitchison and Sil¬
vey (1958, 1960) and Haber (1985a, 1985b). The algorithm used Lagrangian
undetermined multipliers in conjunction with a modified Newton-Raphson
iterative scheme. The modification, which simplifies the method of Haber
(1985a), is to use a simpler matrix than the Hessian matrix. We replace
the Hessian matrix (of the Lagrangian objective function) by its dominant
part, which turns out to be easily inverted. Because the matrices used in the
algorithm proposed in this chapter are very large and must be inverted, this

- 84 -
modification is a very important one. A FORTRAN program ‘mle.restraint’
has been written by the author to implement this modified algorithm.
The asymptotic behavior of the ML estimators computed under the two
sampling schemes—product-Poisson and product-multinomial—was investi¬
gated. The method for deriving the asymptotic distributions represents a
modification to the technique of Aitchison and Silvey (1958). A comparison of
the limiting distributions of the two estimators was made in section 2.4. Some
very interesting results were obtained by studying the asymptotic behavior
in the constraint equation setting. In particular, Theorem 2.4.2 represents
a generalization of the results of Palmgren (1981). The theorem provides a
method for determining when the inferences about the freedom parameters
of a generalized loglinear model of the form dog A/i = X(3 will be invariant
with respect to the sampling assumption. Palmgren (1981) developed some
similar results for the special case when the freedom parameters are part of
a loglinear model.
It is important to note that the asymptotic results are only valid if
the number of populations K is considered fixed and the expected counts
all get large at approximately the same rate. In particular, the asymptotic
arguments do not hold when the covariates are continuous, since the number
of populations (levels of the covariates) can theoretically run off to infinity.
The reason the arguments do not hold is that when we use the method of
Aitchison and Silvey (1958) it is required that the vector n* ldir^n converge
in probability to zero as the total number of observations gets large. This is
the case only when n* = minfni,..., n#} goes to infinity. This drawback

- 85 -
could prove to be temporary. It seems reasonable to assume in many cases,
that as long as the ‘information’ about each parameter is increasing without
bound, the estimators will be consistent and asymptotically normally dis¬
tributed. For example, consider the logistic regression model with continuous
covariates. Although the n^s may all be 1, the ML estimators of the
regression parameters are often consistent and asymptotically normal.
Section 2.5 outlines some miscellaneous results. One result that is
important to the practicing statistician, is that the Lagrange multiplier
statistic is shown to be invariant with respect to choice of the matrix U
(of U'C log An = 0) as long as the model is well defined. An important
implication of this result is that if one fits the model several times, each
time using a different ÍU, matrix, and the Lagrange multiplier statistics
vary more so than can be explained by roundoff, then it could be that the
model is not well defined. Another interesting result is that the Lagrange
multiplier statistic is simply the Pearson chi-squared statistic X2 whenever
the assumptions (Al), (A2), and (A3) are satisfied.
Theoretically the ML fitting algorithm will work for any size problem.
Practically, however, the algorithm is certainly not a model fitting panacea.
The number of parameters that must be estimated gets very large, very fast.
Consider the case where 7 raters rate the same set of objects on a 5 point
scale. Even without covariates, the number of cell probabilities that must be
estimated is 57 = 78,125. It seems the ML fitting method developed in this
chapter is, at least for now, useful for moderate size problems only. It can be
used to analyze longitudinal categorical response data when the number

- 86 -
of measurements taken on each subject is somewhere in the neighborhood of
2 to 6. This is not to take away from the utility of this chapter’s algorithm,
but rather to indicate its breadth of application. In time, with increasing
computer efficiency, much larger data sets may be fitted using this algorithm.

CHAPTER 3
SIMULTANEOUSLY MODELING THE JOINT AND MARGINAL
DISTRIBUTIONS OF MULTIVARIATE POLYTOMOUS
RESPONSE VECTORS
3.1 Introduction
Often times, when given an opportunity to analyze multivariate response
data, the investigator may wish to describe both the joint and marginal
distributions simultaneously. We consider a broad class of models which
imply structure on both the joint and marginal distributions of multivariate
polytomous response vectors. To illustrate the need for such models, we
consider several settings where these models would be useful. For example,
when the multivariate responses represent repeated measures of the same
categorical response across time, one may be interested in how the marginal
distributions are changing across time and how strongly the responses are
associated. The simultaneous investigation of both joint and marginal
distributions is not restricted to the longitudinal data setting. Other examples
include the analysis of rater agreement, cross-over, and social mobility data.
The common thread tying all of these data types together is that the sampling
scheme is such that the different responses are correlated. In longitudinal
studies the same subject responds on several occasions. In rater agreement
studies, raters rate the same objects. In two-period cross-over studies, one
group of subjects receive the two treatments in one order and the other group
receive them in the other order. In social mobility studies, the socio-economic
- 87 -

- 88 -
status of a father-son pair is recorded. When the responses are positively
correlated, these designs result in increased power for detecting differences
between the marginal distributions (Laird, 1991; Zeger, 1988).
This chapter considers the modeling of multivariate categorical responses
in which the same response scale is used for each response. The classes
of models used in this chapter are of the form considered in Chapter 2 of
this dissertation and hence are readily fit using the ML methods of that
chapter. In section 3.2, we give several examples that may be analyzed by
simultaneously modeling the joint and marginal distributions. We introduce
the classes of simultaneous Joint-Marginal models in section 3.3. Several
models are fitted to the data sets of section 3.2.
3.2 Product-Multinomial Sampling Model
Initially, we assume that a random sample of nk subjects is taken from
population k, k = 1,..., K. The number of populations, or covariate profiles,
K is considered to be some fixed integer. The subscript k is allowed to be
compound, i.e. the subscript k is allowed to represent a vector of subscripts
such as
k = [ki,ki,... ,kv).
Suppose that there are T categorical responses Vf1),..., V(T) of interest
and that each response is measured on the same response scale. Let
14 = (V^,..., V£T^)' be the random vector of responses for population k
and Vku, u = l,...,n*. be the nk independent and identically distributed
copies of 14, where Vku denotes the response profile for the uth randomly

- 89 -
chosen person within population k. Notationally we have,
Vku ~ i.i.d. 14, u = 1, — ,n*
For our purposes we can assume that each response takes on values in
{1,2,..., c¿} with probability one. Denote the probability that a randomly
selected subject from population k has response profile i = (¿1,..., iy)' by Trik,
i.e.
P{Vk = =
where i £ {1,..., d} x • • • x {1,..., d}.
The joint distribution of Vk = (V^,..., v£T^)' is specified as {7rifc}. The
marginal distributions of Vk will be denoted by k)}, t = 1,..., T, where
i(t-,k) = P(V^=i), i = l,...,d
Our objective is to model simultaneously the K joint distributions
{TTjjfc}, k = l,...,K
and the KT marginal distributions
{*(<;*)}, i = l,...,T, k = l,...,K.
To help the reader better understand the notation, we consider the one
population bivariate case. When T = 2, the response profiles can be denoted
by i = (¿1,¿2) = (bj)> where i = 1 and j = 1 ,...,d. Since there is
just one population (or covariate profile) the subscript k is always 1 and is
therefore dropped. It follows that {7r¿^} is the joint distribution of (F^1), F(2))'
and {0¿(í)}, t = 1,2 are the two marginal distributions. That is,
= p(vm = i, vw = j), ¿ = 1,j = i,...,d

- 90 -
and
4>¡(t) =
JTi+ = P(V<‘> = i),
*+i = P(W = i),
if Í — 1
if t = 2
for i = 1,2,..d.
Now for each population k, consider the (F x 1 random vector of
indicators
- [V*=*i)’ • • • ’ I(yk=iiT)\
Notice that no information about the Vk is lost since ^ is a one-to-one
function of Vk. Also,
~ ind. Mult(l, {7r¿*.}), k = l,...,K
Therefore, since we have randomly sampled nk subjects from each of the K
populations, we have that for given k
^ki,^k2,---,^knh ~ i.i.d. Mult(l, {7rifc})
and hence the vector
Yk = Y^ ^ku ~ Mult(n*> i^ik})
u=l
is sufficient for the family of distributions {7r¿*.} and {¿(t; k)}.
By independence across populations, the vector vec(Yj,Y2, • • •, Yk) is
sufficient for the joint and marginal distributions of vec(Vj, V2,..., Vk).
Further, the random vector vec(yj, Y2,. • •, Yr-) is product-multinomial, i.e.
Yk = (Ylk,...,YSjty ~ ind Mult(nfc, {7riJb}), k = l,...,K
where 1.,R represent the R = d? different response profiles.

- 91 -
Evidently, Yik represents the number of randomly selected subjects from
population k who have response profile i. That is, the {P**.} represent counts
resulting from a cross-classification of N = Ylk=1 nJfc subjects on T response
variables and a population variable. The data can be displayed in a d? x K
contingency table. By convention, we use lower case Roman letters to denote
realizations of random quantities. For example, yik represents a particular
realization of Yik.
Consider Table 3.1, taken from Hout et al. (1987).
Table 3.1. Interest in Political Campaigns
1960
Not Much
Somewhat
Very Much
Not Much
155
116
64
1956 Somewhat
91
237
171
Very Much
32
91
246
278 444 481
335
499
369
1203
Source: Hout et al. (1987), p. 166, Table 4
Each of 1203 randomly selected subjects was asked in 1956 how inter¬
ested they were in the political campaigns. They responded on the 3-category
ordinal scale: 1 = Not Much, 2 = Somewhat, and 3 = Very Much.
Then, in 1960, each of the subjects was asked the same question and
responded on the same 3-category ordinal scale. Using the above notation,

- 92 -
we let V^1) and V(2) represent the responses in 1956 and 1960. Let i,j =
1,2,3 represent the number of the N = 1203 subjects responding at level
i in 1956 and level j in 1960. Notice that there is just one population
of interest, we drop the population subscript altogether. Finally, for this
bivariate response example, the compound subscript i is replaced by ij. Table
3.1 summarizes the bivariate responses.
As another example, consider the cross-over data of Ezzet and White-
head (1991).
A
Table 3.2. Cross-over Data
B B
1
2
3
4
1
2
3
4
1
59
35
3
2
1
63
40
7
2
2
11
27
2
1
A 2
13
15
2
0
3
0
0
0
0
3
0
0
1
1
4
1
1
0
0
4
0
0
0
0
AB Sequence BA Sequence
(Group 1) (Group 2)
The counts displayed in Table 3.2 are from a study conducted by 3M
Health Care Ltd. to compare the suitability of two inhalation devices (A and
B) in patients who are currently using a standard inhaler device delivering
salbutomal. Two independent groups of subjects participated. Group 1 used
device A for a week followed by device B (sequence AB). Group 2 used the
devices in reverse order (sequence BA).
The response variables V^1) (device A) and V(device B) are ordinal
polytomous. Specifically, they are the self-assessment on clarity of leaflet
instructions accompanying the two devices, recorded on the ordinal four point
scale,

- 93 -
1 = Easy
2 = Only clear after rereading
3 = Not very clear
4 = Confusing.
For this example there are two populations of interest—Group 1 and
Group 2. Let y¡jk represent the number of the nk subjects responding at level
i for device A and level j for device B, where rii = 142 and n2 = 144. Again,
the bivariate response profiles can be denoted by i = ij where i, j = 1, 2, 3,4.
The bivariate responses are summarized in Table 3.2.
3.3 Joint and Marginal Models
Two types of questions that can be posed about Table 3.1 lead to quite
distinct types of models. One question is whether the interest in the political
campaigns was different at the two times. For example, the researcher
may wish to test the hypothesis that there was more interest in the 1960
political campaign than the 1956 political campaign. An investigation into the
marginal distributions is needed to test this hypothesis. For these bivariate
response data, the marginal distributions correspond to the row and column
distributions of Table 3.1. A second question that may be asked is whether
the two responses are associated and if so, how strong is the association. To
answer these questions, we must describe the dependence displayed in the
joint distribution of Table 3.1.
The marginal models we consider will be used to investigate whether
the probability that a randomly selected subject responds at level i or lower
in 1956 is different from the probability that a randomly selected subject
responds at level i or lower in 1960. In this sense, the comparison of marginal

-94-
distributions gives a ‘population averaged’ description of change. That is, we
will describe how the marginal distribution changes on the whole, averaging
over the entire population. In contrast, subject-specific modeling allows us to
investigate how a randomly chosen subject’s response changes from 1956 to
1960. Zeger et al. (1988) discuss at length the difference between population-
average and subject-specific models.
The same types of questions may be posed about the distributions of
Table 3.2. For example, one may wish to determine whether the leaflet
instructions are perceived as clearer for one of the devices. Also, we may
be interested in whether there is a sequence effect. That is, does the order
of ‘exposure’ to the two device’s instruction leaflet affect the perception of
clarity. To answer these two questions we must investigate the marginal
distributions corresponding to the row and column totals of Table 3.2. Finally,
one may be interested in testing whether the association between the two
responses is the same for both sequences. We will consider modeling the joint
distributions to answer this question.
Modeling of marginal distributions is usually conducted separately
from the modeling of joint distributions. We use results from Chapter 2
of this dissertation to show that these models can be fit simultaneously
using maximum likelihood methods. Simultaneously modeling the joint and
marginal distributions leads to several advantages. It will provide a single
test for overall goodness of fit. Also, it provides improved model parsimony,
potentially resulting in better estimates than one would obtain by fitting the
models separately.

- 95 -
We consider four classes of simultaneous models. Let J(S) represent
the class of saturated joint distribution models. These models imply no
structure on the joint distributions and therefore allow for general association
between the T responses. Similarly, let M(5) be the class of marginal
models that assume no structure on the marginal distributions, i.e. M(S)
is the class of saturated marginal models. Denote the classes of unsaturated
models by J{U) and M(U). By simultaneously modeling the joint and
marginal distributions we can consider four classes of models, J(S) n M(S),
J{U) n M(S), J(S) n M(U), and J(U) n M{U). The union of these four
classes will be denoted by J n M. We let the symbol J{M\) n M(M2), where
Mi and M2 are particular models, represent a specific model in J n M. Some
examples of Mi and M2 are Mi — QSY, the quasi-symmetry model, and
M2 = MH, the marginal homogeneity model. The two symbols S and U
will represent either the ‘class’ of saturated and unsaturated models or an
arbitrary model in those classes. The possibility that the joint distribution
structure implied by the joint model J{Mi) will imply that the marginal
distributions are constrained in some way is always there. In this case the
model may not be well defined in the sense of Chapter 2. We address this
issue in section 3.6.
The first class of models J(S) n M(S) is the class of completely
unstructured or fully saturated models. These models fit the data perfectly
and are used primarily for exploratory purposes. If an estimated freedom
parameter is small relative to its standard error, the corresponding effect
may prove to be negligible. In this way, the fit of the saturated model may
suggest simpler models that may fit the data well.

- 96 -
The models in class J(U) n M(S) focus on modeling the joint distri¬
butions. No additional structure on the marginal distribution is assumed.
This class includes ordinary loglinear models for the expected cell frequencies
in the joint distributions. Fitting this simultaneous model is equivalent to
separately fitting the joint distribution model J{JJ) in that the goodness-of-fit
statistic and joint model parameter estimates will be exactly the same. There
is, however, some benefit to fitting the simultaneous model; marginal model
parameter estimates are obtained. In general, these J(U) models are not
designed to estimate effects in marginal distributions. There are exceptions.
For example, the symmetry model for the joint distribution implies that all of
the marginal distributions are equal. Bishop et al. (1975) discuss comparing
the fit of the symmetry (SY) model to the fit of the quasi-symmetry (QSY)
model to test for marginal homogeneity. Our focus will be on models that
do not imply any structure on the marginal distribution. Loglinear models
that assume no relationship among the main effect parameters satisfy this
condition.
The models in class J(S)nM(U) are used to answer questions about the
marginal distributions. They assume no structure for the joint distribution
and hence allow for general association among the responses. Fitting a
J(S) n M{U) model is equivalent to separately fitting the M(U) model in
that the goodness-of-fit statistic and the marginal model parameter estimates
are exactly the same. A simple M(U) model that is often of interest is
the marginal homogeneity (MH) model. Madansky’s (1963) test of marginal
homogeneity is simply the likelihood-ratio test comparing the fit of J(S) n
M(MH) to the saturated model J(S)nilf(5), For bivariate dichotomous

- 97 -
response data, an analogous test using the Lagrange multiplier statistic
(which is shown to be equal to Pearson’s chi-squared statistic in Chapter
2) is McNemar’s (1947) test.
In this chapter, we will focus primarily on the parsimonious models
within the class J(U) n M(U). Often times, a simple model can be found
that fits the data relatively well. Simultaneous inferences about both the
association structure and the marginal distribution structure can be made
using the model or freedom parameter estimates, or goodness-of-fit statistics.
Also, by the parsimony principle, the parameter estimates may be more
reliable than those based on less structured models. See Agresti (1990) and
Bishop et al. (1975) for a discussion of the benefits of using parsimonious
models. We can use models within this class to test such things as MH
given that QSY holds. This can be accomplished by comparing the fit of
J(QSY) n M(MH) to the fit of J(QSY) n M(5). More generally, we may
wish to test for MH given that some simple model Mi holds for the joint
distribution.
Let fj.k = (/zlfc,..., fjLRk)' be the vector of expected frequencies for
population k. That is
Hik — nk^ik
The RK x 1 vector /z is defined as /z = vec(/zi,/Z2,...,^k)- For the marginal
distributions, let {m¿(í; k) = nki(t',k)} represent the marginal distribution
expected cell frequencies. Cumulative marginal probabilities will be denoted
by T/¿(í;fc), i.e.,
t
r]i(t]k) = ^2(f)l/(t-,k), t = l,...,d.
V=\

- 98 -
We consider models in the following classes:
J: CilogAi/x = XiPu ovL1fx = X1f31
M : C2 logA.2/¿ = X2(32 or L2¡j, = ^-2(^2 •
(3.3.1)
The matrices Cj and C2 are either identity, contrast (rows sum to zero),
or zero matrices. The model matrices X\ and X2 are assumed to be of full
column rank. We refer to the parameters in vectors and (32 as freedom
parameters, whereas the components of the parameter vector ¡j, will be called
model parameters.
Evidently, the class of models JnM of (3.3.1) is very broad. Permissible
models for the joint distributions include simple loglinear models as well as
models for log odds ratios using individual cells (e.g. local odds ratios)
or groupings of cells (e.g. global odds ratios which are cross-product
ratios of quadrant probabilities, cf. Dale, 1986). The marginal models of
class M can be loglinear or corresponding logit models (such as adjacent-
categories or baseline-categories logit models) or they can be other types of
multinomial response models, such as cumulative or continuation-ratio logit
models (Agresti, 1990). The second form for each model in (3.3.1) allows for
linear probability or mean response models (Grizzle et al., 1969). All of the
models in JnM can be fit using the methods of Chapter 2. We illustrate the
usefulness of these models by way of example.
3.4 Numerical Examples
Example 1. We begin by simultaneously modeling the joint and marginal
distributions for Table 3.1. Recall that response variable W1) represents a
randomly chosen subject’s response to the political interest question in 1956

- 99 -
and y(2) is a randomly chosen subject’s response to the political interest
question in 1960. Some candidate models for the joint distribution of
(VW, V^2)) include the following:
J(I):
log Vij
— a
+
aY(1)
+
^y(J)
aj
J{QSY) :
log Vij
= a
+
aY(1)
+
V(2)
aY
J(LxL) :
log Vij
= a
+
v(l)
ai
+
y(2)
aj
J(L x L + D) :
log fJ-ij
= a
+
v(l)
ai
+
a]
J(S):
lo gHij
= a
+
ai
+
v(*)
aj
+ cxY¡1)vW, {aY¡1)vW
+ OuiVj
+ SuiVj + 8I(i = j)
+ aY"vm
=ay¡"vW)
where I = independence, QSY = quasi symmetry, L x L — linear-by-linear
association, and L x L + D also adds a main-diagonal parameter. The
latter two models recognize the ordinality of the measurement scale, through
sets of monotone scores {rq} for and {vj} for V^2\ The L x L form of
model fits well when underlying continuous variables have a bivariate normal
distribution (Goodman, 1981; Becker, 1989), and extra parameters for the
main diagonal can account for larger frequencies often observed there when
both dimensions have the same categories.
Candidate models for the marginal distributions of V = (V^1), V(2))
include the following:
M(MH):
log m^t) = (3 + (3? + f3j
M(L x L) :
log rriiit) = ¡3 +(3?+ (3j + (3^vUi
M(CU):
logit T]i(t) =UJi + 7t
M(S):
log mi(t) = (3 + f3f- + /3j + (3j¡v
where CU denotes the cumulative logit and the superscript R is used to
label those parameters related to the ‘level’ of response. There is marginal

- 100 -
homogeneity if there is no association between level of response (R) and
response variable (V) (cf. Agresti, 1989). When the number of levels of
V exceeds two (i.e. T > 2) and V can be considered ordinal, rather than
assume that there are general row effects for levels of V, one could account
for the ordinality by introducing scores for the levels of V. That is, we could
replace PpvUi by fiRVUiVt in the loglinear model and replace 7t by 7vt in the
cumulative logit model. An example where we can consider V as ordinal is
when the T responses represent repeated measures over time. The T levels
of V are then naturally ordered; response at occasion 1 (V^1)), response at
occasion 2 (W2)), ..., response at occasion T (V(T)). For model identifiability,
certain parameters (or more generally, linear combinations of parameters)
were set to zero. For example, the parameter 72 of model M(CU) was set to
zero.
To obtain information about which simultaneous models may fit well,
we first investigate joint and marginal models separately. Table 3.3 contains
likelihood-ratio (G2) and Pearson (AT2) goodness-of-fit statistics for several
models in the class J(U)nM(S). The associated distance or residual degrees
of freedom are listed as well. The linear-by-linear terms used equally spaced
scores for rows and for columns.
Table 3.3. Joint Distribution Models—Goodness of Fit
Model
df
G2
X2
J(S) nM(S)
0
0.00
0.00
J(QSY)nM(S)
1
0.39
0.39
J(LxL + D)nM(S)
2
0.49
0.49
J(Lx L)r M(S)
3
18.58
18.72
J(I) n M(S)
4
245.01
253.09

- 101 -
Both J(QSY) and the simpler J(L x L + D) models fit well. Notice
that the independence model fits poorly as is usually the case for longitudinal
data.
We next fit several models in the class J(S)nM(U). The goodness-of-fit
statistics and the associated residual degrees of freedom for these marginal
models are tabled in Table 3.4.
Table 3.4. Marginal Distribution Models—Goodness of Fit
Model
df
G2
X2
J(S) n M{CU)
1
3.35
3.35
J(S)nM(LxL)
1
4.21
4.20
J(S) n M(MH)
2
38.22
37.49
There is very strong evidence of marginal heterogeneity as measured by the
goodness-of-fit statistic for the model J(S) n M(MH) or as measured by a
comparison of that fit with the fit of some unsaturated model that allows for
marginal heterogeneity.
Finally, we will try to find a good fitting, parsimonious model in the
class J(JJ) n M{U) that simultaneously describes the joint and marginal
distributions. Since the model J(L x L + D) fits the data very well, we
will assume this structure for the joint distribution and simultaneously fit
several candidate marginal models. In section 3.5, we show that the model
J(LxL + D) belongs to a class of joint distribution models that do not imply
any structure on the marginal distribution. It therefore follows that residual
degrees of freedom for the simultaneous model J(L x L 4- D) n M(U) can be

- 102 -
computed as follows,
dfies[J{L xL + D) n M(U)} = dfies[J(L xL + D)] + dfTes[M(U)].
This follows since the model is well defined in the sense of Chapter 2 and since,
for well defined models, residual degrees of freedom is simply the difference
between the number of constraints implied by the simpler model and the
number of constraints implied by the less structured model. Table 3.5 contains
the result of fitting several models in the class J(L x L + D) n M(U).
Table 3.5. Candidate Models in J(L x L + D) n M{U)—Goodness of Fit
Model
J{Lx L +D)nM(S)
J(L x L +D)nM(CU)
J{LxL +D)nM{LxL)
J(L x L +D)nM(MH)
df
(P
JP
2
0.49
0.49
3
3.84
3.82
3
4.68
4.66
4
38.73
38.15
The simple model J(L x L + D) n M(CU) fits the data very well
(G2 = 3.84, df = 3). This model implies that the joint and marginal
distributions simultaneously follow the models,
J(L x L + D) : logfJ'tj = a + aYw + ajw +9-ij + 8I(i = j)
M(CU) : logit T)i(t) = ují + 7i
In Table 3.6, we give the ML estimates of the freedom parameters for
this model along with their corresponding estimate of standard errors.

- 103 -
Table 3.6. Estimates of Freedom Parameters for
Model J(L x L + D) n M(CU)
Parameter
Estimate
Std. Error
a
0.085
0.662
v(i)
ai
2.430
0.349
a.T
1.605
0.203
v(3)
<
1.606
0.325
y(3)
«2
1.172
0.192
e
0.563
0.081
8
0.355
0.084
U>1
-1.255
0.063
U>2
0.435
0.057
7i
0.341
0.058
To test for marginal homogeneity in the context of this model, we can use
either of two asymptotically equivalent x2(l) test statistics:
G2 = 38.73 - 3.84 = 34.89
w2 / 0.341,2 04 5-7
W '-0.058''
where W2 is the squared Wald statistic. The P-values for both of these tests
are less than 0.001. We conclude that there is strong evidence of marginal
heterogeneity. We need not, and should not, stop here. Since we are working
with model and freedom parameters, we can continue with other model-based
inferences. Interval estimation of certain interesting freedom parameters is
considered next.
The interpretation of the parameter 71 is as follows: The odds that a
randomly selected subject would have responded at level i or less in 1956 is
exp(71) times higher than the odds that a randomly selected subject would
have responded at level i or less in 1960. Thus, the freedom parameter 71
measures the departure from marginal homogeneity in that the two odds are

- 104 -
identical if and only if ^ = 0. We use the delta method to compute a 95%
confidence interval for the odds ratio exp(7i); it is [ 1.324 , 1.488 ]. Thus,
based on the data at hand, we estimate that the odds that a subject would
respond at level i or less in 1956 is between 1.324 and 1.488 times higher
than the odds that a subject would respond at level i or less in 1960. There
is significant evidence of increased political interest in 1960 relative to 1956.
Next we consider the association between the two responses. The estimated
odds that the response in 1960 was ‘very much’ instead of ‘somewhat’ is
exp(0 + 2¿) = 3.57 times higher when the response in 1956 was ‘very much’
than when it was ‘somewhat’. The same estimated odds ratio applies when
the response was ‘somewhat’ instead of ‘not much’. Similarly, the estimated
odds that the response in 1960 was ‘very much’ instead of ‘not much’ is
exp(40 + 28) = 19.34 times higher when the response in 1956 was ‘very much’
than when in was ‘not much’. In summary, there is evidence of strong positive
association between the response in 1956 and the response in 1960 and there
is evidence that there was greater political interest in 1960 than in 1956.
Suppose we ignored the fact that the same subjects responded to the
political interest question in 1956 and 1960. If we treated the two responses as
independent, then the row and column marginal counts would be distributed
as independent multinomials with the same index N =â–  1203 and probability
vectors {0¿(l)} and {^(2)}. Then it follows that separately fitting the
marginal model M{U) under this independence assumption is equivalent
to fitting the simultaneous model J(7) n M(U). By results of Liang and
Zeger (1986), the estimates of parameters in M{U) would be consistent, even
when the responses are not truly independent. However, the estimates of

- 105 -
the corresponding standard errors would no longer be valid. One way to see
that we are losing information by incorrectly assuming independence is by
comparing the likelihood-ratio statistic for testing MH assuming J(I) holds
to the likelihood-ratio statistic for testing MH assuming that J(L x L + D)
holds. The former is G2 = 268.33-247.74 = 20.59 and the latter is G2 = 34.89.
Both of these values would be compared to a tabled %2(l) value. Evidently,
by accounting for the dependence between the responses we have greater
evidence of marginal heterogeneity. Another way of illustrating the effect
of wrongly assuming independence between the responses is by looking at
the freedom parameter estimates and their estimated standard errors for
different models. Table 3.7 contains estimates of 7j and the corresponding
standard error estimate under three different models of interest. Notice that
the standard errors are similar when one used either the saturated or the
diagonal parameter model for the joint distribution.
Table 3.7. Freedom Parameter Estimates and Standard Errors
Model
df
7i
se(7i)
J(S) n M(CU)
1
0.342
0.058
J(L x L + D) n M(CU)
3
0.341
0.058
J(I) rM(CU)
5
0.343
0.076
We have shown that there may be problems with assuming too much
structure on the joint distribution; for example, unreasonably assuming
independence. Similarly, we should be concerned with assuming too little
structure on the joint distribution. In this case, too many freedom parameters
require estimation and the overall fit may be unreliable. A good model is one
that fits the data at hand relatively well and is robust to the white noise

- 106 -
present in the data generation. That is, a good fitting model with model
parameter estimates that change very little for different realizations of the
random data vector, is considered a good model. For example, the saturated
model fits perfectly but has parameter estimates that may change greatly
for different realizations. In this sense the saturated model may not be a
good one; it may be unreliable. When we ignore the association structure by
separately fitting marginal models, we are tacitly using the saturated model
for the joint distribution. Table 3.8 illustrates why we should search for a
good fitting, parsimonious model. Note that the standard errors of expected
cell frequency estimates are inflated when we assume a saturated model for
the joint distribution. The more parsimonious model J(L x L + D) nM(CU)
fits as well as the less structured model J(S)r\M(CU), yet it is more reliable
in the sense described above.
Table 3.8. Estimated Cell Means and Standard Errors
for Models J(S) n M(CU) and J{LxL + D) n M(CU)
J(S) n
M(CU)
J(L x
L + D)nM(CU)
A¿i
se(fia)
Ait
se(im)
152.79
11.49
154.28
10.56
127.00
8.80
123.08
6.56
64.87
7.82
66.98
7.16
82.74
7.53
83.25
4.89
237.30
13.80
237.30
13.80
159.53
9.95
159.00
8.12
31.41
5.52
29.37
4.10
99.14
8.44
103.05
6.28
248.23
13.98
246.70
13.16

- 107 —
Example 2. We continue with the cross-over data example of section 3.2.
Denote the set of 18 local odds ratios by {r^}, where
7~ij k
and 7represents the probability that a randomly chosen subject from
Group (G) k responds at the ith level for device A (Vl1)) and the jth level for
device B (W2)). Recall that cumulative marginal probabilities are denoted
by
u\-i EUi *V+fc, if t = 1 (device A)
lEUi7r+i/*> if t = 2 (device B)
where i = 1,2,3,4 and k = 1,2. To elucidate, 773(2; 1) represents the
probability that a randomly chosen subject from Group 1 will respond at
level 3 or lower for device B (V(2)).
Some possible models for the joint distributions of (v£1\ V^)', k = 1,2
include the following:
AS):
J{VWG,VWG,VWVW):
J(LxL):
J(VWG,VWG):
J(VW,VW,G):
J(UA(G)):
J{UA):
log Hijk = aijk
log fMijk = a + ocYW + ctJ(J)
+«?+«rG+<,G+ log** =a + ar“> + ajm + a° +
+a%')G + avt',vi,)ui Vj
log ** = a + <*rW + <’> + af + + aJ">e
log** = a + a,1"'' + + of
l°g g* =u + 0k
log rijk=u
where J(S) is a fully saturated model, J{V^G, V^G, V(1)y(2)) assumes
no three-factor interaction, J(L x L) implies that there is no three-factor

- 108 -
interaction and that the association between the ordinal responses can
be accounted for by including a linear-by-linear association parameter,
is the mutual independence model, and J(V^G, V^G)
implies that and W2) are conditionally independent given G. The model
J(UA(G)) implies uniform association within levels of G, and J(UA) is the
simple model that assumes this uniform association is the same for both levels
of G. When the row and column scores {u¿} and {vj} are equally spaced,
models J(L x L) and J(UA) are equivalent. It is shown in section 3.6 that
model V"(2), G) implies that the marginal distributions of (VF), F(2))'
do not depend on G. When this happens, the simultaneous model will be ill
defined whenever the marginal model constrains the marginal distributions to
be equal across levels of G. We will not consider this particular model for this
reason. The rest of the models do not imply any structure on the marginal
distributions. Also, notice that simultaneously fitting J(V^G, V^G) and
some marginal model M(U) is equivalent to separately fitting M(U) when
the row and column marginal counts are treated as independent multinomials
within each level of G.
The marginal models we fitted include the following cumulative logit
models:
M(S): logit(r/,(i; k)) = fiut
M(VG): logit k)) = ft + ft + ft0 + Pla
M(V, G): logit(T?,(<; fc)) = ft + ft + ftG
M(V): logit (>,.((;&)) = ft+ftV
M( 1): logit(iji(<; fc)) = ft

- 109 -
where M(5) is the saturated model and M(VG) is the proportional-odds
cumulative-logit model for the marginal probabilities that allows for otherwise
general association between the response variable V, the group or population
variable G, and the response ‘level’ R. In the literature on cross-over designs,
a second-order interaction among V, G, and R is said to be a ‘carryover’ effect.
The model M(V, G) implies that there is no second-order interaction among
the variables V, G, and R, i.e. the model implies that there is no carryover
effect. The model M(V) implies that there is no G effect, i.e. no sequence
effect. Finally, the simple model M( 1) implies that there is no V or G effect.
To make these models identifiable, we place the following restrictions on the
freedom parameters.
PX = -PX = Pv
P? = -PX = PG
qvg _ f PVG, if t + k = 3
^tk \ 0, otherwise
With this parameterization, f3v, (3G, and (3VG measure device, sequence, and
carryover effects, respectively.
Table 3.9 displays the goodness-of-fit statistics and their associated
degrees of freedom for several simultaneous models. The L x L model used
the equally spaced row and column scores = ¿} and {Vj = j}.

- 110 -
Table 3.9. Cross-over Data Models—Goodness of Fit
Model
df
G2
X2
J{S)nM(S)
0
0
0
J(S)nM(VG)
6
10.55
6.91
J(UA)nM(S)
17
17.36
9.19
J(V^G, V^G, V^VW) n M(VG)
15
14.28
10.65
J(L x L) nM(VG)
23
28.52
27.00
J(VWG,VWG) n M(VG)
24
37.92
58.77
J(UA(G)) r M(VG)
22
28.45
26.11
J{UA) n M(VG)
23
28.52
27.00
J(UA)nM(V,G)
24
29.97
29.64
J(UA)nM{V)
25
31.05
30.32
J(UA) n M(l)
26
70.51
64.87
Evidently the parsimonious model J(UA)n M[V) fits the data very well.
This model implies that there is no period or carryover effect and that the
uniform association structure is the same for each sequence group. There is
evidence of a significant device effect (G2 = 70.51 - 31.05 = 39.46, df — 1).
We will proceed to describe this device effect. The freedom parameter ML
estimates and their corresponding standard error estimates are tabled in Table
3.10.
Table 3.10. Freedom Parameter ML Estimates
for Model J(UA)nM(V)
Parameter
Estimate
St Error
0.469
0.148
Pi
0.542
0.096
P2
3.189
0.219
Pz
4.360
0.375
PV
0.511
0.082
These estimates also indicate that there is a significant device effect; the
Wald statistic which is based on 1 degree of freedom takes on the value of

- Ill -
W2 = (^e0v))2 = 38.8. The magnitude of the device effect can be estimated
using /3V. Specifically, the odds of responding j + 1 or higher for device B is
estimated to be e2^v = 2.78 times higher than the odds for device A. Using
the delta method, an approximate 95% confidence interval for this odds ratio
is (1.87, 3.69). Since the higher responses correspond to less perceived clarity
of the instructional leaflet, we conclude that there is evidence suggesting a
significant improvement of device A over device B in terms of perceived clarity
of instructions. We can describe the association between the two responses
using to. For either sequence group, the odds of responding at level i instead
of i + l for device A is estimated to be exp(0.469) = 1.6 times higher when the
response for device B was i rather than i +1. This holds for each i and j. In
summary, there is a moderate positive association between the two responses,
the strength of association being the same for both sequence groups. There
also is significant evidence of increased perceived clarity for device A over
device B.
3.5. Product-Multinomial Versus Product-Poisson
Estimators: An Application
In this section and in section 3.6, we explore some of the more practical
aspects of model fitting for categorical data. In this section we will illustrate,
by way of example, how to determine when inferences based on freedom
parameters will be the same under both sampling assumptions—product-
multinomial and product-Poisson. The method of determination is a direct
consequence of Theorem 2.4.2. In section 3.6, we address, at least partially,

- 112 -
the issue of whether or not the model is well defined. Closely related to this
is the computation of residual and model degrees of freedom.
Consider the data taken from the Harvard Study of Air Pollution and
Health. The data, displayed in Table 3.11, can be found in Agresti (1990,
p.414); they were supplied by Dr. James Ware.
Table 3.11. Children’s Respiratory Illness Data
No Maternal
Child’s Respiratory Illness Smoking
Age 7Age 8Age 9 Age 10
No
Yes
No No No
237
10
Yes
15
4
Yes No
16
2
Yes
7
3
Yes No No
24
3
Yes
3
2
Yes No
6
2
Yes
5
11
Maternal
Smoking
Age 10
No
Yes
118
6
8
2
11
1
6
4
7
3
3
1
4
2
4
7
Source: Agresti (1990, p.414), supplied by Dr. James Ware
The two groups of children—those with smoking mothers and those with
nonsmoking mothers—were followed for four years, from age 7 to age 10. At
each occasion, each child was tested for respiratory illness. The response
vector for the kth (k = 1,2) group of children is Vk = (V^*\ V^),
where response Vjf* is binary; either the disease is present or it is not. Our
goal is to find a parsimonious, simultaneous model that fits the data well.
Using this model, we will be able to address questions such as “is mother’s
smoking status associated with the child’s respiratory illness status” or “are
the odds of having respiratory illness the same for all four years?”

- 113 -
After fitting several simultaneous models, we finally settled on the
following good-fitting (G2 = 14.33, df = 22) simultaneous model.
J : logmju, = a + aVm + a]"’ + <■' + cY"‘ + a? + al"’s + c)
+ alms + *rs + aif,v<” + + aY<‘>vm
y(2)
4- a
y(2)y(s)
jk
+ ajl
y(*)y(*) . y(»)y(^)
+ OL
kl
M : logit ; s) = 6 + 6Y,
(3.5.1)
(3.5.2)
where 6Y satisfies the following,
0Y=0V = $V = $V; 9\ = 0
This model ((3.5.1) n (3.5.2)) implies that there are no three-factor
interactions among the five factors—the four responses and the covariate,
there is no significant group (Smoker) effect, and that there is marginal
homogeneity among the first three times. There is an indication that the
odds of having respiratory illness are lower when the child is 10 years old. In
fact, the test statistic value used for testing marginal homogeneity across all
four times was significantly large (G2 = 24.29 - 14.33 = 9.96, df = l).
Our objective in this section is to determine which of the freedom
parameter estimates, if any, are affected by assuming the counts are product-
Poisson rather than product-multinomial. We will use Theorem 2.4.2. To
invoke the results of that theorem more directly, we will rewrite the model
using the matrix notation of Chapter 2. The model can be written as
Glog = X(3,

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o o o
o o o
rH ^D rH
rH O
o o
rH ^D rH ^D rH (O rH ^D rH ^D rH ^D rH ^D rH
rH rH ^D
rH rH rH rH CZ2
rH O
O O
O
O
O
rH rH ^D
rH rH rH rH rH rH rH rH ^D ^D ^D
O
O rH
i—l O
o o
rH O O
rH O rH
rH i—I O
rH O t—I
rH O
O O
O O
O rH
i—I t-H O O rH i—I
O
rH O O
O O O
o o o
o
rH O O
o o o
o o o
o o
o o
o o
o o
o o
O i—I
rH rH C3
rH O
o o
o o
o o
o o
o o
O rH
O rH
i—I i—i i—I i—I o
o o o
o o o
o o o
O rH O
o o
o o o
o o o
o o o
o o o
o o o
o o o
O rH O
rH O O
o o o
o o
O rH
rH O
o o
o o
o o
o o
o o
o o
o o
o o
O rH
rH O
rH rH rH rH rH rH rH rH ^D ^D ^D
O rH
O O
O O
O o
o o
o o
o o
o o
o o
o o
o o
O T 1
O rH
rH O
o o
o o o
o o o
o o o
o o o
o o o
o o o
o o o
o o o
o o o
o o o
o o o
O rH O
rH O O
o o o
o o o
1—I rH rH o rH rH rH O
V,
i?
n
s
c/}
N
s
v>
'S-S
$
a
e
r—I
s
y(i)y(2) y(i)yW y(i)y(<) y(*)y(s) y(*)y(*) y(*)y(<)

- 116 -
and
$2 = (M'7-
Also, the vector of expected cell counts n is a 2 • 24 x 1 vector and is
defined as
H — (/¿mili/¿11121 j • • • j /¿22221, ^11112 > • • • 5/^22222)'•
That is, the last subscript (corresponding to the sth group) is changing the
slowest and the other 4 subscripts are in lexicographical order.
In view of Theorem 2.4.2, we must determine, for i = 1,2, whether or
not C{ is a contrast matrix. If it is not, then we must find those columns of
that span a set containing the range space of 02lm¡, where qi — m\ — 16 and
g2 = 4 m2. Recall that qi is the number of response functions within each
independent population for the ith model. For example, for this data set, the
second model (i = 2), which is the marginal model (3.5.2), has g2 = 4 logits to
be modeled within each of the two population groups (children with smoking
mothers and children with nonsmoking mothers). As in the statement of the
theorem, we will find a minimal spanning subset.
Since matrix C\ is not a contrast matrix, we wish to find the columns
of Xi that span a space containing the range space of ©21i6. With the
parameterization we have used, we can easily see that the first and the
sixth columns of Xi span the required space. Also, C2 is a contrast matrix.
Therefore, it follows by Theorem 2.4.2 that the two asymptotic variances of
the freedom parameter estimators, computed under the two different sampling
assumptions, are related as follows,
varCP{M)) = var(^(p)) " ( ^21 A22 ) ’

- 117 -
where Anisal6xl6 matrix with zeroes everywhere except in rows 1 and 6
and columns 1 and 6 and all the other A‘J,s are zero matrices.
Table 3.12 displays the freedom parameter estimators and their es¬
timated standard errors, which were calculated under the two sampling
assumptions. Notice that only those standard errors corresponding to the
parameters a and af are different for the two sampling schemes. These are
the parameters that correspond to the first and sixth columns of X\.
Table 3.12. Product-Multinomial versus Product-Poisson
Freedom Parameter Estimation
Parameter
Estimate
Product-Multinomial
Standard Error
Product-Poisson
Standard Error
a
1.67
0.216
0.228
„ y(i)
«1
-1.20
0.304
0.304
v(»)
al
-1.35
0.342
0.342
af3)
-1.07
0.266
0.266
af4)
-0.39
0.288
0.288
of
0.63
0.000
0.091
oy(1)5
0.00
0.000
0.000
av(J)s
U11
0.00
0.000
0.000
fy
all
0.00
0.000
0.000
av(4)s
U11
0.00
0.000
0.000
V(1)V(2)
clH
0.73
0.323
0.323
V(i)y(3)
U11
1.30
0.303
0.303
nVWvM
** 11
1.64
0.321
0.321
y(2)v(3)
1.56
0.304
0.304
y(2)V(4)
0.98
0.327
0.327
V(s)y(4)
uii
0.92
0.226
0.226
e
2.02
0.134
0.134
9V
-0.38
0.126
0.126
One last remark worth mentioning is with regard to the standard error
estimates of the estimated expected cell counts {Ai/¡fcz¿}- The precision

- 118 -
estimates will be different for the two sampling schemes. In fact, the
relationship (2.4.6), viz.
-(P)-(P)'
var(/¿(M)) = var(/¿(p)) - ®f ————,
ni
allows us to determine how different the two variances will be. For example,
the estimated expected cell count for cell (1,1,1,1,1) is /¿mil — 232.80
and the standard errors are 7.029 and 14.292 corresponding to the product-
multinomial and product-Poisson sampling assumptions. The difference in
standard errors is substantial. In contrast, the estimated expected cell count
for cell (1,1,2,2,1) is /j1122i = 4.09 and the two standard errors are 1.324 and
1.342. The product-Poisson standard error estimate is only slightly inflated.
Suppose that, instead of assuming the logit model (3.5.2) for the
marginal parameters, we used the equivalent loglinear model. That is, we
will modify the matrices C2, A2 and X2, and the vector /32, so that the logit
model is equivalently expressed as a loglinear model. Let = ®jI&, A% = A2
(no modification is necessary for this example), and
*2 =
(1 1
1 0
1 1
1 0
1 1
1 0
1 1
1 0
1 1
1 0
1 1
1 0
1 1
1 0
1 1
VI 0
1 0
1 0
0 1
0 1
0 0
0 0
0 0
0 0
1 0
1 0
0 1
0 1
0 0
0 0
0 0
0 0
0 1
0 1
0 0
0 0
1 0
1 0
0 0
0 0
0 0
0 0
0 0
0 0
1 0
1 0
0 0
0 0
0 0
0 0
1 0
1 0
0 1
0 1
0 0
0 0
0 0
0 0
0 0
0 0
0 0
0 0
0 0
0 0
1 !\
0 1
1 1
0 1
1 1
0 1
0 1
0 1
1 0
0 0
1 0
0 0
1 0
0 0
0 0
0 0/

- 119 -
With this specification, the logit model is equivalent to the loglinear
model
M : logm¿(í; k) = A + Af + Af + Afs + Af v + Af, (3.5.3)
where Af v satisfies
and {m,(t;A;)} is the set of expected marginal counts. That is, =
nk(f)i{t\k). The vector /32 is thus defined as
ft = (A, Af, Af^Af, Aâ„¢, Aâ„¢, Aâ„¢, Af^, Af)'
Notice that the loglinear model (3.5.3) includes the VS effect. This
effect must be included so that the model is well defined. We will discuss this
further in the next section, section 3.6.
The matrix C2 is not a contrast matrix for the loglinear representation
of the marginal model. Therefore, to determine which freedom parameter
estimators are unaffected by the sampling assumption, we must find, among
the columns of X2, the minimal spanning set for At(©flmj) = At(©^l8).
Notice that the number of response functions, within each population, for
the marginal model is now m2 = q% =8, not q2 = 4 as it was for the logit
model. Again, with the parameterization we have chosen, we can easily see
that the first and tenth columns of X2 span a set that contains the range space
of ®jl8. Invoking Theorem 2.4.2, we have the following result. Letting the
vector ¡3 represent the freedom parameter vector for model ((3.5.1)n(3.5.3)),
var(/?(M>) = var(/3(p>) - A = var(/3
- 120 -
where the elements of the partitioned matrix A are
A;; = /o, if (M)¿{i,6}x{i,6>
Ai; = / 0, if {k,l) ( {1,6} x {1,10}
kl \ + 0, otherwise
A2i = /0, if (fc, Z)*{1,10} X {1,6}
kl \ ^ 0, otherwise
and
A22= io, if (k,l)4 {1,10} X {1,10}
kl \ > 0, otherwise.
By expressing (3 = vec^, (32) as (3 = (^i,^2j-"->^26)S we can state the
result in another way: If (i,j) £ {1,6,17,26} x {1,6,17,26} then cov(0¿, 6j)
is the same under both sampling assumptions. If (i,j) is in the set then the
covariances may be different.
To illustrate, we compare the standard errors for the loglinear parameter
estimators. It happens that all of the freedom parameter estimators are the
same (see Theorem 2.4.1) and all of the standard errors are the same except
those associated with the liJ, 6th, 17th, and 26th parameters, namely a, af, A,
and Af. For these four, the standard error estimates were related as follows
se(o:|Poisson) = se(d|multinomial) + 0.012
se(df | Poisson) = se(df |multinomial) + 0.091
se(A|Poisson) = se(A|multinomial) + 0.016
se(Af |Poisson) = se(Af ¡multinomial) +0.091.

- 121 -
In summary, we were able to easily determine when inferences using
certain freedom parameter estimators would be the same under both sampling
schemes. This holds for a very broad class of generalized loglinear models of
the form ClogA/i = X/3. Basically, if the matrix C is a contrast matrix,
that is both C\ and C2 are contrast matrices, all of the inferences are the
same. On the other hand, if, for example, C¿ of C is an identity matrix
then we must look at the design matrix Xi to determine which columns form
a minimal spanning subset for the range space of some matrix of the form
©fTm¡. When Ci is an identity matrix, is the number of response
functions, within each population (or level of covariate), that are modeled via
CilogAi/i = Xi/3i.
3.6 Well-Defined Models and the Computation of
Residual Degrees of Freedom
We made some remarks above with regard to models being well or ill de¬
fined. To illustrate, we use the simple example in which the joint distribution
model is J(SY) and the marginal distribution model is We stated
that the model J(SY) n M(MH) is ill defined since the constraints implied
by the symmetry model J(SY), namely that the marginal distributions are
equal, are the model constraints of We will show that, for the
one population setting, as long as the main-effects loglinear parameters are
allowed to be arbitrary (up to freedom parameter identifiability constraints)
the joint distribution model will only imply that the expected marginal counts
satisfy the (multinomial) identifiability constraints. In all other respects the
expected marginal counts are allowed to be arbitrary positive numbers. That

- 122 -
is, the joint distribution model and the marginal distribution model will
not include redundant constraints and the simultaneous model will be well
defined. For this example, J(SY) restricts the main-effects parameters to
satisfy
rm>.
Evidently, the sufficient condition for the model to be generally well defined
is not met. We also discuss sufficient conditions for a simultaneous model to
be well defined when there are covariates present.
A simultaneous model will necessarily be well defined if the following
three conditions hold: The joint distribution model must be well defined. The
marginal distribution model must be well defined. And, the joint distribution
model must only constrain the expected marginal counts to satisfy the
identifiability constraints. The first two conditions hold whenever the models
do not contain redundant and/or conflicting constraints; the identifiability
constraints being included. For example if one covariate is present, as long
as the generalized loglinear portion of the model allows for a perfect fit to
the sums of expected counts within each level of the covariate, the model will
be well defined. In what follows we consider the two response, one covariate
case to illustrate how one can identify a large class of simultaneous models
that will be well defined. The extension to arbitrary numbers of responses
and covariates is straightforward.
Suppose that A and B are two response variables. We will initially
allow the number of response categories for A and B, namely I and J, to
be different. Since this chapter deals with situations when the responses

- 123 -
are measured on the same scale (i.e. I = J), we will also address the
sufficient conditions for model well definedness in that case. Denote the K
level covariate by P. The following lemma identifies a large class of joint
distribution models that only imply that the expected marginal counts satisfy
the identifiability constraints. It is important to point out that we will be
referring to two types of identifiability constraints. ‘Identifiability’ constraints
are those constraints associated with multinomial sampling, namely that cer¬
tain sums of probabilities add up to 1. ‘Freedom identifiability’ constraints are
those constraints that are necessary to ensure that each freedom parameter in
the model is estimable. The identifiability constraints for /z will generically be
labelled as iderzi(/z) in this section. Similarly, let the identifiability constraints
for m, the vector of expected marginal counts, be denoted by ident{m). These
constraints are implied by ident(n).
Lemma 3.6.1. Let the hierarchical loglinear model (AP,BP) be specified as
either
log/z = X*/3*, ident(pL), or U*' log/z = 0, ident^pi).
Suppose that the joint distribution model [0j] can be specified as either
log p = X/3, ident(p), or U'\ogp = 0, ident(p).
If [© j] is no more restrictive than (AP,BP) in the sense that
M(X)DM(X*) or M(U)CM(U*),
then [0j] only constrains the expected marginal counts to satisfy the identifi¬
ability constraints ident{rn).

- 124 -
Proof: Write the model (AP,BP) as
log Vijk = a + oc£ + a f + af + afkp + afkp,
where without loss of generality the freedom identifiability constraints are
ap = a? = af = off = a?p = afp = afp = 0, V», j, k,
and the identifiability constraints ident(fx) are
k = l,...,K.
* i
Using the identifiability constraints we can write
nk = exp(a + oip)'yk'yk , k = l,...,K,
where
Hence,
ik = +atkp)
i=l
Ik = 5>xp(a? + afkp)-
j=l
a + ak = logâ„¢* - logTfc1 - logTjf
Now all of the freedom parameters not constrained by the freedom identifi¬
ability constraints or the identifiability constraints are completely arbitrary.
It follows that (7^}and (7¿f}, which are functions of these arbitrary freedom
parameters, are also completely arbitrary.
Therefore,
j
mi(l,k) =
3=1
= exp(lognfc - log 7^ - log7jf + af + afkp)-yf
= exp (log nk - log 7^ + ocf + afkp)
_ nk exp(q4 + afp)
Ik

- 125 -
That is, this set of expected marginal counts follows a saturated multinomial
loglinear model. Similarly,
(n - Uk eXP(af +afkP) â–  _ J i -i rr
mj(2j k) g , J k
"4
follow a saturated multinomial loglinear model. Since the two sets of expected
marginal counts are functions of different arbitrary parameters we have that
the entire set of expected marginal counts are constrained only to satisfy the
identifiability constraints zdent(m), viz.
i J
m¿(l, &) — nki and k) — nfc, k = l,...,K.
t=i j=i
Now, if any joint distribution model is less restrictive in the sense stated in the
lemma, it must be that the model must only constrain the expected marginal
counts to satisfy ident(m). This is what we set out to show. g
As a special case, suppose that the covariate P has just one level,
i.e. K = 1. Lemma 3.6.1 tells us that a sufficient condition for the joint
distribution model to only constrain the expected marginal counts to satisfy
ident(m) is that the main-effects parameters {o^} and {a?} be arbitrary up
to the freedom identifiability constraints. In fact, for the case I = J, in view
of the proof of the lemma, if we constrained the main-effects parameters to
satisfy
af = af, * =
then expected marginal counts would be constrained to satisfy marginal
homogeneity. Another generalization of Lemma 3.6.1 involves the situation
when there is more than one covariate. If there was more than one covariate,
say P and Q, then the joint distribution model should be no more restrictive

- 126 -
than the hierarchical loglinear model (APQ, BPQ) for the conclusion of
Lemma 3.6.1 to hold.
Since most reasonable joint distribution models will be well defined we
assume this to be the case and hence are left to show that the marginal
distribution model is well defined. To show this, we simply must show
that the generalized loglinear or linear marginal model constraints and
the identifiability constraints ident(m) (which are implied by ident(//)) are
independent. We will initially assume that I need not equal J. Let the factors
Ri and R2 represent the level of response to factors A and B. That is, R\
is an I level factor and R2 is a J level factor. A simple loglinear model for
the expected marginal counts can be written as ((P1? P), (P2> P)). What this
means is that the expected marginal counts satisfy
logm¡(l,fc) = /3' +P?' + 0lr, ¿ = 1k = l,...,K
logmJ(2,fc)=iai+/3f +PlP, j = = (3.6.1)
/3Pl = flf2 = PlP = P\p — 0, ident(m).
Suppose now that I — J. As before, let the factor R represent the
common levels of response for both response factors A and B. Also, the
factor V will again be defined to be the response variable factor. For this
example, V is a two-level factor taking on the values 1, corresponding to
the ‘first’ response A, and 2, corresponding to the ‘second’ response B. For
longitudinal data, V is referred to as the ‘Occasion’ variable. Since I — J we
can consider an even simpler model. We could assume that
/9f' = = T*,
* = 1 ,â– â– â– ,!

- 127 -
and consider the model (P, VP), which can be specified as
logm¿(í, k) = r + rfi + rp + rf + Tt^p, t = 1,2, i = 1,...,/, fc = 1,.. .,K,
(3.6.2)
where
r tY — /?*, i = 1,2
rk + TtkP = Ptpi ¿ = i,2, fc = i,...,ür,
the r parameters satisfy the freedom constraints,
tv -tp - tvp - tkp - 0 vf fr
and the identifiability constraints ident(m) are satisfied. Notice that the
model (P, VP) only makes sense when I = J; it implies marginal homogeneity
of the A and B response distributions. The following lemma provides us with
a way of identifying a large class of marginal distribution models that are well
defined. It is concerned with the case when I need not equal J. Lemma 3.6.3
applies when I — J. Each of these lemmas is easily generalizable to situations
when there are many response variables and many covariates.
Lemma 3.6.2 Suppose that the marginal distribution model ((Pi, P), (P2, P))
can be written as either
\ogm = X*(3*, ident(m) or U*‘ logm = 0, ident(m),
where ident{m) are those identifiability constraints implied by ident{n).
Specify the marginal distribution model [0m] as
log m = X(3, ident{rn) or U1 log m = 0, ident(m).

- 128 -
If [0M] is no more restrictive than ((Ri, P), (P2> P)) in the sense that
M{X) 2 M(X*) or M{U) C M{U*)
then [Ojtf] is well defined.
Proof: By equation (3.6.1), the marginal model {{R\, P), (R2, P)), without
the identifiability constraints, implies that
*(M) = X777^1’*1) = +#bP)XexP(#Rl) and
j=l *=1
s(2,k) = x mi(2’= exP (P2 + z5^) X exp(/3f1 )•
i=l 3=1
Hence, the s(t,A:), which are functions of 2 * K arbitrary parameters, are
arbitrary. Since the identifiability constraints ident(m) constrain the s(i, k) to
satisfy s(i, k) = nk, k = 1,..., K, t = 1,2 and the model constraints allow the
s(t, k) to be completely arbitrary, it follows that the model ((Ri, P), (R2, P))
is well defined. Also, any less restrictive marginal distribution model will also
be well defined. _
Notice that in the proof of Lemma 3.6.2 the conclusion would still hold
if the sums X)f=i exp(Pfl) and exp(Pf2) were constrained to equal each
other. This will be important when we show that the model (R, VP) is well
defined.
Suppose now that I = J so that the model (R,VP) is reasonable. This
next lemma identifies a large class of marginal distribution models that are
well defined when the responses are measured on the same scale.

- 129 -
Lemma 3.6.3 Suppose that the model (R,VP) can be written as either
logm = X*/3*, ident(rn) or U*' logm = 0, ident{m).
Specify the marginal distribution model [0^] as
logm = X/3, ident{m) or U' logm = 0, ident{rn).
If [0M] is no more restrictive than (R,VP) in the sense that
M(X) D M(X*) or M(U) C M{U*)
then it is well defined.
Proof: By equation (3.6.2), we can write the sums s(t,k) = as
s(t, k) = exp(r + rtv + rf + t%p) £ exp(r/*).
i
Notice that the first exponential term is completely arbitrary; it is a function
of 2 * K independent parameters. Therefore the set of sums (s(i,fc)} is not
constrained in any way by the model constraints, logm = X*(3*. As in the
proof of Lemma 3.6.2, it follows that the marginal distribution model (R, VP)
is well defined. Finally, any less restrictive model will also be well defined, g
In view of the proof of Lemma 3.6.3, the model (R, V, P) would not be
well defined; neither would (RV,P). In order for the marginal distribution
model to be well defined the loglinear model must include the VP effect. We
can easily generalize the results of Lemma 3.6.3. Suppose that there are two
covariates, say P and Q. It can be shown that any marginal distribution
model that is no more restrictive than the loglinear model (R,VPQ) is well
defined. A marginal distribution model that is specified as a cumulative- or

- 130 -
adjacent categories-logit model would be well defined if the model allows the
sums {s(t,k)} to be completely arbitrary.
We now state an important theorem that addresses the issue of model
well definedness. The theorem is specifically for the case when the response
variables A and B are measured on the same scale and there is just one
covariate P. It can easily be generalized to the case of several distinct
responses and several covariates.
Theorem 3.6.1 Suppose that the joint distribution model [0j] is no
more restrictive than the loglinear model (AP,BP) and that the marginal
distribution model [O^f] is no more restrictive than the loglinear model
(R,VP). It follows that the simultaneous model [0j n &m] 15 weM defined.
Proof: The proof follows immediately by Lemmas 3.6.1 and 3.6.3 and
the fact that a simultaneous model is well defined if the following conditions
hold: Both the joint and marginal distribution models are well defined and
the joint distribution model only constrains the expected marginal counts to
satisfy the identifiability constraints ident(rn). g
A few remarks about Theorem 3.6.1 are in order. Firstly, when there is
only one population of interest the sufficient condition is that the main-effects
parameters are allowed to be arbitrary. It follows that such models as quasi
symmetry (J(QSY)) satisfy these sufficient conditions. Also, models such as
J{UA(G)) and J(UA) of the cross-over example satisfy the conditions. This
follows since the model J(UA) is equivalent to the model J(L x L) which
satisfies the sufficient conditions of the theorem; it is less restrictive than
(VWG,VWG).

- 131 -
For the example of section 3.5, we see that had we left the effect VS
out of the marginal loglinear model (3.5.3), the marginal model would have
constrained the sums {s(t,k)} to lie in some restricted space. This can be
seen by noting that
s(tik) = Y exP (P + P? + Pt +Pk+ *üV)
t=i
= exp {P+PY + Pi) Y exp (A*v + Pi1)
i
and that neither exp(/3+/3ty +¡3%) or exp(A-*y+/?/*) is completely arbitrary;
s(t,k) is constrained to satisfy s(t,k) = Ktpk for some Kt and pk. Therefore,
the marginal model constraints and the identifiability constraints are not
independent. That is, model ((3.5.1) n (3.5.3)) would not be well defined if
the effect VS were not included in (3.5.3). This also follows directly from
Theorem 3.6.1. Using the program ‘mle.restraint’, an attempt was made to
fit the ill-defined model. The algorithm did not converge. In practice, this
nonconvergence could very well indicate that the model is ill defined (see
section 2.5) as it did in this example.
If a simultaneous model is well defined it follows that the residual degrees
of freedom can be computed as
dfrea[®J H ©Af] ~ dfrea[Qj\ + d/rei[©Af]- (3.6.3)
since the model constraints are nonredundant. For example, the residual
degrees of freedom for measuring goodness of fit of the simultaneous model
J(L x L + D) n M{U) used in the political interest data example can be
computed in this way. This follows since the model J(L x L + D) satisfies
the sufficient conditions of Theorem 3.6.1 and so, if M{U) is well defined, the

- 132 -
simultaneous model J(L x L + D) n M(U) is well defined. In contrast, the
model G) used for the cross-over data example, does not satisfy
the conditions of the theorem since the effects V^G and V^G are omitted.
In fact, the model implies that there is no Group (G) by Response level
(R) association. Therefore, the simultaneous model comprised of this joint
distribution model along with the marginal cumulative-logit model M(V) is
ill defined since M(V) implies the same constraints. Equation (3.6.3) does
not apply in this case.
3.7 Discussion
In this chapter, we introduced a broad class of models that imply struc¬
ture on both the joint and marginal distributions of multivariate categorical
response vectors when the response scale was the same for each response. We
showed that these models can be fit using the ML fitting method of Chapter
2. Several numerical examples were considered, illustrating the usefulness
of simultaneously modeling the joint and marginal distributions. All of the
models were fitted using the FORTRAN program ‘mle.restraint’, which was
developed by the author.
Model parsimony was the impetus behind this entire chapter. Our
objective was to find parsimonious models that both fit the data well and
provided us with straightforward interpretations of freedom parameters. The
models often included parameters that measured departures from indepen¬
dence among the responses, as well as parameters that measured departure
from marginal homogeneity. It was shown, via a numerical example, that
parsimonious modeling may result in more efficient and reliable estimation

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of both model and freedom parameters, the researcher must find a balance
between a model that is too structured and one that is not structured enough.
The author fully intends to conduct simulation studies to better understand
the importance of parsimonious modeling in this setting.
Although we provide somewhat general results regarding compatibility
of the joint and marginal models, there still is a need for more general results.
We discuss the case when the joint and marginal models can be expressed,
at least equivalently, as loglinear models. More general results are needed for
other types of models, such as cumulative-logit and linear models. For these
simultaneous models to be useful to the practitioner, a general method to
determine whether the constraints implied by the two models are independent
must be developed. The proposition in section 3.6 is a step in the right
direction.
A factor that could impede the use of this method to fit models to very
large data sets is the input requirements. The algorithm requires a substantial
amount of input. For example, consider the input required for the example
in section 3.5. The matrices C, A, and X all must be input. Although the
required input is simple to determine, there is much energy expended inputing
the information. An input program must be developed and implemented in
the program ‘mle.restraint’.
The assessment of model goodness of fit is straightforward when using
the ML method. The (log) likelihood-ratio statistic G2, the Pearson statistic
X2, or the Wald statistic W2 can be used for this purpose. Of interest to
the practicing statistician, is the ability to assess how far wrong you can
be by assuming that the responses are independent. The test statistic used

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for this purpose is simply the likelihood-ratio statistic that measures how
‘far apart’ the models J(J) n M(U) and J(S) n M(U) are. Because the
model J(I) n M(U) is nested within the model J(S) n M(I7), one can use,
as a measure of this distance, the difference between the two likelihood-ratio
statistics, viz. G2[J{I) nM(U)] - G2[J(S) nM(t/)]. More generally, there are
many assumptions one can make about the association structure among the
responses. With the methods of this dissertation, one can easily derive tests
for the validity of the assumptions.
As an alternative to longitudinal type sampling designs, a cross-sectional
sample may be taken. Cross-sectional sampling involves sampling indepen¬
dent groups of subjects for each response. The research questions posed about
the marginal distributions are such that they could by answered using cross-
sectional data. In this sense, the marginal models are ‘population averaged’
models (Zeger et ah, 1988). However, a cross-sectional sampling design
results in more subject variability, since nonhomogeneous subjects are used for
each response, and the detection of differences in the marginal distributions
may be clouded by these subject effects (Laird, 1991). Further, with cross-
sectional studies, we are unable to explore the association structure among
the responses. This information, regarding the association structure, may be
of substantive importance in some situations.

CHAPTER 4
LOGLINEAR MODEL FITTING WITH INCOMPLETE DATA
4.1 Introduction
We consider making inferences about loglinear model parameters when
only disjoint sums of the complete data are observed. Inferences will be made
based on the maximum likelihood estimates of the model parameters and an
estimate of precision of these estimates. As an example, consider the data in
Table 1 of Goodman (1974). Each of 216 respondents was classified as being
universalistic or particularistic when confronted by each of four situations
(A, B, C, D) of role conflict. Goodman (1974) postulated the presence of an
underlying two-level latent factor W which was not observed. Within a level
of the latent factor the manifest variables (A, B, C, D) are assumed to be
mutually independent. Thus, the latent class structure would allow us to
simply explain the relationship among the four manifest variables. In this
setting the unobservable complete data are the counts resulting from a cross¬
classification on the four manifest factors and the latent factor. The data,
if observable, could be displayed in a 25 contingency table. The observable
incomplete data are the counts obtained by summing over the two levels of
the latent factor, i.e. the incomplete data are disjoint sums of the complete
data. As in Goodman (1974), we assume the complete data means follow a
loglinear model which implies conditional independence among the manifest
factors (A, B, C, D) given the latent factor W. Our objectives include finding
- 135 -

- 136 -
the maximum likelihood estimates of the loglinear parameters based on
the observed data, estimating their precision, computing other model based
estimators and their standard errors, and testing model goodness of fit.
There are many ways to find the maximum likelihood estimators, each
method having its positive and negative features. For example, we could work
directly with the incomplete-data likelihood, which is usually complicated
relative to the complete-data likelihood, and use a Newton-Raphson or Fisher¬
scoring algorithm. Palmgren and Ekholm (1987) and Haberman (1989) use
these methods to obtain maximum likelihood estimates and their standard
errors. We could avoid the complicated likelihood altogether and use the
Expectation-Maximization algorithm (Dempster et al., 1977). Sundberg
(1976) discusses the properties of the EM algorithm when it is used to fit
models to data coming from the regular exponential family. In section 4.2
the EM algorithm is explored in greater detail.
Unlike the other approaches, the EM algorithm is insensitive to starting
values. This is important in practice since we seldom have any idea what
a reasonable starting value is. Another positive feature, not shared by the
other methods, is that the convergence to the maximum is monotonic, i.e.
the likelihood is increased at each successive iteration. Drawbacks to the EM
algorithm are that (1) it is relatively slow and (2) an estimate of precision
of the parameter estimate is not obtained as a by-product of the algorithm.
N-R and Fisher-scoring, on the other hand, are faster and, as a by-product,
provide us with an estimate of precision. The slow convergence of the EM
algorithm can be mitigated somewhat using the acceleration methods of
Meilijson (1989) or Louis (1982). Also, increased computer efficiency has

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made the slow convergence less of an issue. In section 4.3.2 we address
the second drawback of the EM algorithm by deriving an explicit form for
the observed information matrix when the complete data are independent
Poissons with means following a loglinear model. The observed information
matrix is computed upon convergence of the EM algorithm and then inverted.
The inverse will serve as the estimate of precision. In section 4.5 we explore
an iterative scheme that uses both NR and EM, exploiting each of their strong
points.
4.2 Review of the EM Algorithm
The EM algorithm is generally used in those estimation problems in
which the likelihood is complicated, rendering it difficult or impractical to
maximize, but in which the data can be viewed as being some function
of complete data which, had they been observed, evaluation of maximum
likelihood estimates would be simple. Unlike many other statistical root¬
finding algorithms, the EM algorithm does not require explicit calculation of
the score vector or its derivative. It uses much simpler functions.
The EM algorithm is by no means a new method for finding maximum
likelihood estimates. Goodman (1974) essentially used it. Sundberg (1976)
discusses it at length when used in the exponential family case. Dempster,
Laird, and Rubin (1977) provide us with a review of the method as well as
some of its properties. Subsequent work with the EM-algorithm has been
primarily devoted to improving the speed of its convergence (Louis, 1982;
Meilijson, 1989).

- 138 -
4.2.1 General Results
Suppose the complete data X has density fx(x; 9) with respect to some
measure. Let Y — Y(X), a function of the complete data, denote the observed
data. It follows that the density of Y is
/y(y;0)= / fx{x-,0)dv(x), (4.2.1)
Jr
where R — {x : Y(a;) = y} and v is some appropriate measure. Since Y is a
function of X, the joint density of X and Y can be written as
fx,r(x> y;e) = 9) â–  IR(x).
Hence, the conditional density of X given Y = y is
t (r.„ üi _ fx,YÍx fxr{x’y’9> ~ fr(y,g) ~ fY(y,e)
Therefore, the log likelihood based on Y is
(4.2.2)
M^;y) = l°gfY{y,0) = logfx(x;9) - log fxlY(x;y,9).
Taking the conditional expectation (given Y = y) at 0O gives us
M0;v) = -E(M»;#)iv = i/,«.)
= E(ex(0; X)\Y = y, »„) - E(ix[Y(e, y; X)\Y = y, $„)
= «(Mo,»)-ff(Mo,y).
The EM algorithm is defined by
Q(9(m+1\9(m\y) = rnaxQ(y,y(m),y), (4.2.3)
0
i.e. given the mth iterate estimate of 9, 9(m\ the next iterate is that value of
9 that maximizes Q(9,9(m\y).

- 139 -
The following properties of the EM algorithm are verified in the
appendix. The proofs follow from Dempster et al. (1977) and Louis (1982).
In what follows S denotes a score vector and I an information matrix.
Property 1:
If and are the mth and m+ l4t iterate estimates obtained via
the EM algorithm then
M0(m+1);s/)>M<'(ra);!/),
i.e. the log likelihood is increased at each successive iteration.
Property 2:
The sequence of EM iterates {9^m\m > 1} satisfy, whenever 9
converges to 0(°°) as ra -> oo,
I*-) = *M0(oo);y) = °-
i.e. the estimates converge to a zero of the score vector for Y.
Property 3:
For any 90,
§g[Q(«,«o,y) I».] = sy(g 0;y) = E(sx(ff0;X)iv =
Property 4:
For any 90,
Jy(9„;y) = E(Ix(S„-X)\Y = - var(Sx(«„;X)|r = y,60).
Briefly, property 1 implies that the incomplete-data likelihood is in¬
creased with each successive iteration, property 2 says that the EM algorithm

- 140 -
can be used to find a zero of the incomplete-data score function, property 3
provides us with a way of evaluating the score function (see Meilijson, 1989),
and finally property 4 gives us an expression for the observed information
matrix based on the incomplete data. These four properties of the EM
algorithm will be explored in detail in the next section which deals with
the special case in which the complete data have distribution in the regular
exponential family.
4.2.2 Exponential Family Results
The exponential families of distributions play an important role in statis¬
tical inference. Many data generating mechanisms can be modeled assuming
that the underlying distribution is a member of the regular exponential family.
In this section we consider properties of the regular exponential family that
are relevant to the use of the EM algorithm. Specifically, we will make use
of the results of this section, which are due primarily to Sundberg (1974), to
justify results for Poisson loglinear models with missing data.
Let the complete data vector X have density, with respect to some
measure, in the regular exponential family. That is assume that
fxfaP) = a(x)exp(T'(x)/3 - c(p)), (4.2.4)
where T(x) = (Ti(x),T2(x), .. .,Tp(x))' and /3 is a canonical parameter vector
of length p. Let X — (x : fx(x\P) > 0}.
Some well known properties of the regular exponential family include
1. T(X) is sufficient for (3
2' H§L = Ei>(nX)) and
w§=™Anx)).
(4.2.5)
3.

- 141 -
These properties of (4.2.5) are shown in Lehmann (1983, pp. 29,30). The
properties follow immediately upon repeated differentiation of Jx fx{x',fi)dy,(x)
with respect to (3. Lehmann (1983) showed that the derivative could be passed
through the integral.
Suppose that the incomplete data vector Y is a (many to one) function
of X, i.e. Y = Y(X). For notational convenience, we let t = T(x) and Ir(x)
represent the indicator of membership in R = {x : Y(x) = y}. It follows by
equation (4.2.2) that
/jr|y(*;y>£)
Íx(x]^)-Ir(x) _ a(x) exp(t'(3 - c((3)) • IR(x)
/y(y;/3) ¡Ra(x)exP(t'0 - c(/3))dv(x)
= a(x) exp(t'(3 - c*(/5; y)) â–  IR(x) = a*(x) exp(t'fi - c*(/3; y)),
(4.2.6)
where a*(cc) = ci(x)-Ir(x) and c*(/3;y) = log JRa(x) exp(t'(3)du(x). Hence the
conditional distribution of X given Y — y is also a member of the exponential
family (Sundberg, 1974). Again by properties of the exponential family we
have
1. - Ed(T(X)\Y = y) and
2- = var^T(x)lF = »)•
Using (4.2.2) and (4.2.6) we can reexpress the density of Y as
/y(y;/3)
_ fx{x;(3)-IR(x)
fx\y{x;y,P)
_ a(x) exp(t'(3 - c({3)) • IR(x)
a(x) exp(t'/3 - c*(/3; y)) • IR(x)
= exp(c*(/3; y) - c((3))

- 142
Our objective is to maximize /y(y; /3) with respect to (3. Or, equivalently,
we are to maximize the log likelihood
M/3;y) = c*(/3;y)-c(/?)
(4.2.8)
with respect to (3.
For well behaved £Y((3]y) we can find the value of (3, say /5, that
maximizes it by solving the score equations
c (s-v)--f ac*(/3;y) dc(/3)
sY{p,y) d(3^Y{p,y) d/3 dp o.
(4.2.9)
Notice that by properties given in (4.2.5) and (4.2.7), this is equivalent to
solving the equation
Sr(P;y) = MT(X)\Y = y) - E„(T{X)) = 0. (4.2.10)
There are many ways to solve (4.2.10). One possibility is to use the following
iterative scheme:
(1) Find EeM(T(X)\Y = y)
(2) Solve for /3(-+» in E^(T(X)) = EpM(T(X)\Y = y) (4.2.11)
(3) If ||/3(*d - /5^I/+1^| > TOL then replace (3^ by /3^+1^ and go to (1).
Else stop.
We show in Appendix B that the iterative scheme (4.2.11) is simply the EM
algorithm. The convergence properties are discussed in Sundberg (1976).
One important note with regard to the EM algorithm (4.2.11) is that if
£y(/3;y) is not so well behaved, e.g. the score vector 5y(/5;y) has multiple
roots some of which may be associated with a minimum, then the particular

- 143 -
solution /3, obtained via the EM algorithm, will be a local maximum likelihood
estimate. This follows since the likelihood increases monotonically with each
successive EM iteration.
Upon convergence of the algorithm, we can use the negative Hessian
matrix evaluated at /3 to estimate the observed information matrix based on
the incomplete data. The negative Hessian is
Mfcy) =-QfojjtYifcy)
d2c((3) d2c*(p)
dp'dp dP'dp
= var,(T(X))-var,(T(X)|y = y)
(4.2.12)
=lx^y)-lx\Y{P\y)
This expression for the negative Hessian was noted by Sundberg (1974).
He referred to the matrix IX\y as a measure of information loss. With
regard to lost information, let us suppose the observed data Y are such
that T(X) = g(Y). That is, the sufficient statistic for P is a function of
Y. Intuitively we would expect no loss of information since we are able to
observe the sufficient statistic and hence we expect IX\Y to be identically the
zero matrix. In fact, this follows since T(x) is constant on R = {x : Y(cc) = y}
whenever T(x) = g(y). Hence c*(/3;y) = exp(t'/3) JRa(x)dv(x) which is linear
in p. Thus
Jx|y(/5;y)
d2c*(p;y)
dpdp
In view of equation (4.2.9), instead of using the iterative scheme de¬
scribed in (4.2.11), we could work directly with the incomplete data likelihood
¿y(P‘i V) and implement a Newton-Raphson or Fisher-scoring algorithm to find
a root to the nonlinear equation. The program NLIN described in Appendix B

- 144 -
can be used to this end. Notice that both Sy (/3; y) and IY(/?; y) (or a numerical
approximation thereof) would need to be computed at each iteration.
Specifically, the iterative scheme can be written as
(1) Compute /3(I/+1) = /3(u) + (AY((3("); y))~l SY{(3^\y)
(2) If H/Jbd _ /5(l'+1)|| > TOL then replace by /5(I/+1) and go to (1).
Else stop. (4.2.13)
where AY(/3; y) = IY((3; y) if the Newton-Raphson method is used, AY(/3] y) =
Ep(IY(¡3\ Y)) if the Fisher-scoring method is used, or AY((3; y) is a numerical
approximation to the observed or expected information. See section 4.5 for
details on the approximation method.
In section 4.5, we consider an iterative scheme that is a modifica¬
tion/combination of the two schemes (4.2.11) and (4.2.13). The modified
algorithm for solving (4.2.10) exploits the virtues of both these iterative
schemes.
4.3 Loglinear Model Fitting with Incomplete Data
We investigate more closely the special case of incomplete Poisson data
with means following a loglinear model. The assumption that the complete
data are distributed as product Poisson, i.e. the components are independent
Poisson random variables, is not as restrictive as it seems. We use results
of Birch (1963) and Palmgren (1981) to show that maximum likelihood
inferences about the parameters that are not fixed by sample design are the
same whether the data are product Poisson or multinomial. To this end,
we derive an expression for the variance of the multinomial cell probability

- 145 -
estimates when the model parameters are estimated under the product
Poisson assumption.
Section 4.3.1 shows that the EM algorithm takes on a particularly simple
form when the complete data are assumed to be product Poisson with means
following a loglinear model. In section 4.3.2 we derive an explicit formula for
the observed information matrix that is based on the observable incomplete
data. Section 4.3.3 discusses inferences for multinomial loglinear models.
4.3.1 The EM Algorithm for Poisson Loglinear Models
Let X = (Xi, X2,..., X„) represent the “complete” data vector of cell
counts and suppose that
Xi ~ indep. Poisson(/i¿), i = l,2, ...,n
where pi = /i¿(/5) satisfies the loglinear model logp(/3) = Z/3. Here Z is some
nxp full rank model matrix and ¡3 is a p x 1 parameter vector.
Suppose only certain disjoint sums of X are observable. Let Y =
{Y\,Y2,... ,Ym) = LX denote the observable (or “incomplete”) data. Here
L is an m x n matrix (m < n) that satisfies the following three properties:
(1) Each element is a ‘O’ or a ‘1’
(2) There is at most one ‘1’ per column (4.3.1)
(3) There is at least one ‘1’ per row
Properties (1) and (2) of (4.3.1) ensure that the components of Y
are independent Poisson random variables while property (3) precludes a
noninformative row of zeroes.

- 146 -
Denote realizations of X and Y by x and y. The objective of this section
is to find the maximum likelihood estimate of /?, denoted by /3, based on the
observed data. Writing the density of the complete data X as
fx(x;(3) = a(x) -exp(x'Zf3 - l'eZ/3) (4.3.2)
we see that fx has form (4.2.4) and that a sufficient statistic for (3 is Z'X. It
follows by (4.2.8) that Y = LX has log likelihood of the form
M/3;y) = c*(/3;y)-c(/?). (4.3.3)
where c* and c are functions defined in section 4.2.2. But, by properties of the
matrix L, we know that Y has a product Poisson distribution. Specifically,
Yi ~ ind Poisson(T'/x), i =
where L\ is the ith row of L and p. is the vector of complete data means. Since
the complete data means are a function of some model parameters through
log(/¿) = Z(3, we have that = L'iexp(Z/3). It is important to note that
log(L'ifi) is generally a nonlinear function of (3. For this reason, the model
fitting is somewhat more complicated.
Using the fact that Y is product Poisson, we have that the log likelihood
of Y is
m m
M/3; y) = Y, y* los(3 exP (ZP)) -Y,1'* exp(zP) + Hv) (4.3.4)
1 1
where the function h{y) is independent of the parameter (3. Now, we
differentiate equation (4.3.4) with respect to ¡3 to obtain an expression for

- 147 -
the score vector. It is
o m m
y)=? T^mz'DML'' - Z'DM ?Li
m m
= Z,£l(,í)( S “ Z'D{L< ~ !») - Z'-D(M)1»
m m
= Z'DM S ¿;exp(^j?)£i) + ^Wf1- - £ L-) -
= MW £ ()Li + Z'/x - ZV
(4.3.5)
where in the last line and ‘—’ represent componentwise operators. As
shown in section 4.2.2. equation (4.2.10), the log likelihood of the incomplete
data can alternatively be expressed as
V) = Ee(Z'X\Y = y)~ Eg(Z'X)
since dc*(/3)/d/3 = Ep(Z'X\Y — y) and <9c(/3)/<9/3 = Ep(Z'X). Evidently,
since Ep(Z'X) = Z'fi, it must be that
E,(Z'X\Y = y) = Z'[n- (1. - Vlm + £’(#))] (4.3.6)
Therefore, the EM algorithm is simply
(1) Find Z'K/JM) ■ (1„ - L'lm + ¿'(j^))]
(2) Solve for /?(■'+» in Z>(/3<-+’>) = Z'\p(f3<->) • (1. - L-lm + ¿'(j^))]
(3) If ||/3^^ - /3(,/+1)|| > TOL then replace /3b7) by /3(,/+1) and go to (1).
Else stop. (4.3.7)
In practice, finding a reasonable starting value for /3, say (3^°\ is very
difficult. However, in view of the first step of the EM algorithm, we need
only be concerned with an initial estimate of ¿i. Notice that if /A°), the initial

- 148 -
guess for /¿, satisfies L/i(°) = y then we have tacitly chosen an appropriate /?(°)
to start the algorithm. This is so since we can go to step (2) of the algorithm
and calculate ¡3^ the solution to the equation. In fact,
/3W = (Z'Z)-1Z' log//0).
Thus, the EM algorithm has the nice feature that, not only is it
insensitive to starting values, but also reasonable starting values are simple
to find. A FORTRAN program ‘em.loglin’ has been written to actually
implement the EM algorithm as defined in (4.3.7).
4.3.2 Obtaining the Observed Information Matrix
In the previous section we showed how one can obtain maximum likeli¬
hood estimates of the loglinear model parameters using the EM algorithm. In
this section we address the major drawback of the EM algorithm; an estimate
of the precision of these ML estimates is not obtained as a by-product of the
algorithm. We derive an explicit formula for the observed information matrix
associated with the loglinear model parameters that is intuitively appealing
and simple to evaluate. Upon convergence of the EM algorithm the observed
information matrix is evaluated at the ML estimates and inverted. The
inverse information can be used as an estimate of precision (Agresti, 1990).
Notice that in this section we consider using the observed information rather
than the expected information. We follow the lead of Efron and Hinckley
(1978) which builds a case for the preferred use of the observed information.
If desired, however, the expected information can easily be computed since
the observed information is shown to be a linear function of the incomplete
data.

- 149 -
Recall the setup in the previous section. Only disjoint sums of a complete
data vector X, which is product Poisson, are observable. The complete data
means are assumed to follow a loglinear model of the form log ¡j, = Z/3. By
expression (4.2.12) of section 4.2.2. we see that the observed information
matrix based on the incomplete data has form
IriPw) = varp{Z'X) - yaxp{Z'X\Y = y)
— Ix(fl) - (Adjustment Matrix)
This expression is intuitively appealing since varp(Z'X) = Z'D(fi)Z is the
expected (and observed) information for ¡3 treating the complete data, X, as
if it were observed, while var/3(Z'X|F = y) is an adjustment that is necessary
because we do not actually observe X but only LX = Y. The amount of
information lost by observing only Y is determined by the conditional variance
of the sufficient statistic Z'X given LX = y.
At this point, one could derive a formula for the adjustment matrix as
in a technical report by the author. The gist of the argument was that the
distribution of X\Y — y has a simple form when Y represents disjoint sums of
the independent Poisson random variables X and so the conditional variance
of X (or Z'X) given Y = y can easily be computed. A main result of that
technical report was that
cov(Xa,X6| LX = y)
\ Ma • J(r(a)=0) + &(«) JT (l jT ) ’ ^(r(a)>0) | * -^(a=6)
1 r(a)^ r(a)r >
+ { Vr(a) LT^ ' J(r(a)=r(6)) | * ^(a#6)
where 1^ is the indicator function and r(j) is defined as follows:
(4.3.8),
row number
0,
in which ‘1’ occurs, for the jth column of L,
if a ‘1’ does not occur in column j of L.

- 150 -
In this dissertation we will take a different approach. The explicit
form of the score statistic for Y was derived in equation (4.3.5). Since the
observed information is nothing but the negative Hessian of the log likelihood,
we can obtain an explicit formula for the observed information by simply
differentiating the negative of the score function with respect to /3'. The
appendix shows how one arrives at
lY(P\y) = —£p¡SY(P\y)
= Z'D^)(±J^LiL't)mZ
1 « (4-3.9)
(m t i \ \ /
= Z'D(u)L'D(j^)LD(lx)Z - Z'D(V(*^))D(riZ,
where the ‘—’ in the last line represents componentwise division.
Notice that the expected information matrix has a particularly simple
form, viz.
E,(IrV)-,Y)) = Z'DM(jr7^¡LiLl)D(riZ
= Z< D^)L' D-\L^)LD{ii)Z.
(4.3.10)
Using either of the results in (4.3.8) or (4.3.9), we derive an explicit form
for the observed information matrix for several examples.

- 151 -
Example 1: Missing Components—When certain components are
unobservable, L will be an identity matrix with rows missing. It follows
that the observed information matrix is
M/3;!/) = Z'DMZ - Z'D{Mn)Z
where M is a diagonal matrix with jth diagonal element (M)jj = 1 - I(r(j) >
0).
Example 2: Latent Class Models—Suppose that counts resulting from
a cross-classification on several factors are observable and that classification
on an additional K-level latent factor is unobservable. We let the subscript i
represent a compound subscript identifying classification on observable factors
while the subscript j indexes the K latent classes. Denote the complete
data vector of cell counts by X = (Xu,..., Xuc, ..., Xm\,..., Xmx;)T =
{Xij} and the incomplete data by Y = {X¿+}. Notice that Y = LX
where L = 1^ ® Im. One possible latent class model assumes the means
of the unobservable complete data, say /i¿y, follow a loglinear model that
implies conditional independence of observed factors given the latent factor
classification (Haberman, 1979). It follows that the observed information
matrix is
M/3; y) = Z'D(n)Z - Z\ ©¡I, V¡)Z (4.3.11)
where each is the covariance of a K x 1 multinomial vector with index
yi = Xi+ and cell probabilities {/¿¿j/ Y,f=i V-iji j = 1» • • ■ j X}.
Example 3: Partially Classified Data Models—Consider the two factor
nonignorable nonresponse model with one supplemental margin (Little &
Rubin, section 11.6, 1987). The complete data X are counts resulting from

- 152 -
a cross-classification on two factors and F2, along with a dichotomous
nonresponse indicator R. Suppose the Fy classification is always known and
that R indicates whether or not the F2 classification is known. To make
inferences about the classification probabilities and missing data assumptions,
Little & Rubin assume the complete data means follow a loglinear model.
Variance estimates of the loglinear parameters are easily derived since the
observed data have form Y = LX and L satisfies (4.3.1).
4.3.3 Inferences for Multinomial Loglinear Models
Previously, we assumed that the complete data were distributed as
product Poisson, i.e. the complete data components are independent Poisson
random variables. However, the sample size is often fixed by design so that
the distribution of the complete data vector may really be multinomial. This
follows since a product Poisson vector given the total is multinomial. Since
the total sample size is considered a random variable when the product
Poisson assumption is used, the assumption seems to be unreasonable.
Fortunately, Birch (1963) and Palmgren (1981) showed that maximum
likelihood inferences about all of the loglinear parameters that are not fixed by
design are the same whether one assumes the distribution is product Poisson
or multinomial. Therefore, it is general practice to assume the data are
product Poisson since the Poisson distribution is in the regular exponential
family and has an unconstrained canonical parameter. The Poisson loglinear
model is an example of a generalized linear model (McCullagh and Nelder,
1989) which makes it simple to work with.

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In this section we discuss making inferences about loglinear parameters
when the sampling design is such that the total sample size is considered
fixed but the data are not completely observed, i.e. there is missing data. It
is not obvious that the results of Birch extend to the case of incomplete data.
Therefore, we provide a detailed discussion of the extension to the missing
data case.
The Setup. In the following argument we assume that the matrix L is such
that each column has at least one ‘1’ in it. This requirement results in the
incomplete data Y = LX having the same sum total as the complete data,
i.e. VmY = 1 'mLX = l'nXd=N. We also require the loglinear model to include
an intercept term. This intercept term will be the parameter that is fixed by
design, since the total sample size N will be considered fixed.
Full-Multinomial Sampling. Suppose that the complete data vector X has
a multinomial distribution, i.e.
X = (Xlt...,Xny ~ Mult(iV,7r(0)),
where N = l'nX is the fixed total sample size and 7r(8) = (7^(0),... ,7rn(0))'
represents the vector of cell probabilities that satisfy X)"=i 7r¿(^) — 1- Since N
is considered fixed, it makes sense to write the cell means as /i¿(0) = iWj(0)
so that Z)"=i/¿*(0) = N. Assume also that the cell means {/¿¿(0)} follow the
loglinear model
log m(9) = oc + x'i/3, t = l,...,n,
where is a p x 1 vector and 8 = (a,/3contains the so called loglinear
parameters.

- 154 -
Further, suppose that only Y = (Yi,..., Ym)' = LX is observable. The
matrix L, which is of dimension mxn (m < n) will be required to satisfy the
3 conditions of (4.3.1) as well as
(4) L has at least one T’ in each column.
It follows that
Y = (Y1,...,Ym)' ~ Mult(iV,Ltt(9)),
where Ltt(9) = (.Lj7r(0), ..., Z/m7r(0))'. Again expressing the cell means as
rj{9) = Lfj,(9) = NLtt(9), we have that the incomplete data cell means satisfy
r]i(9) = L-p(0) = L\ exp(aln + X/3) = exp(a)L¿ exp(X/3).
Also, since there is a constraint on the /¿¿(#) there is a constraint on the r¡i{9).
In fact, the r]i satisfy 2 r¡i{9) = L\)n{9) = 1 'np(0) = N. Also, the log
means satisfy \ogrji(9) = logexp(a)L¿ exp(X/3) = a + log(L¿ exp(X¡3).
Denote the model parameter space for the multinomial scenario by 0^
and notice that
0«={*=w ')'=
i
Evidently the set Qm is constrained and so 0^f is not equal to the (p + 1)
dimensional real space.
Consider the one-to-one transformation 9^9* = (t,/3where r =
Y,™ p¿(0). It follows that under this new parameterization the rji satisfy
m
log r7¿(<9*) = logr-log(]TL;exp(A73)) + log exp(X/3)
i
,r = = Z^(exp(a)Líexp(x^))
i i
m
=> a = logr - log(^) L\ exp(X/3))
i
since

- 155 -
We will call the new parameter space Q*M and note that it is
0Ji = {0* = (t, 13’)': T = N,f3eR
The incomplete-data likelihood under the (M)ultinomial assumption can
be written in terms of this new parameterization as
4MV*;s/) = Eviiog^*) - iviogjv
i
= ^y¿(logr-log(^L; exp(X/3)) + logL'¿ exp(X/3)) - NlogiV
i i
= N logr - N log N + £ Vi log L'i e*PixP) - N l°g(£ L'i exp(X/3)
i i
= &L'i exp(X0) - Nlog^L'iCxpiXP)) = t2(P),
(4.3.12)
since V0* g 0^, r = N. Therefore, the incomplete-data multinomial log
likelihood is independent of r. Also, since the parameter (3 is free of
constraints, we can maximize y) with respect to 9* by simply setting
f = N and maximizing the unconstrained function i2(/3) with respect to (3.
In this context we refer to a as being fixed by sample design since it is a
function of the other parameters (3 and the fixed sample size N.
Product-Poisson Sampling. In contrast to the first sampling scheme, the
total sample size is not considered fixed. Assume that the complete data
X = (Xi,..., An)' are distributed as product Poisson, i.e.
X{ ~ ind Poisson(/¿¿(0)), z = l,...,n,
where the parameter 9 is unconstrained and the means satisfy
log Hi(9) = a + x'fl, i =
Again, we assume that the complete data are not observable and that
we only are able to see Y = (Yi,...,ym) = LX with L satisfying the same

- 156 -
four properties that it did in the multinomial setup. The vector Y is then
distributed as product Poisson. Specifically,
Yi ~ ind Poisson(L¿/¿(0) = 7^(0)), i = l,...,m.
The cell means ?7¿(0) satisfy the model
logffc(0) = a + log(L-exp(X/3)), i= l,...,m.
or using the same reparameterization [6 6*) as above
log»7i(^*) = logr - log (L'i exp{Xf3)) + log {L\ exp(X(3)).
i
We will denote the model parameter space for the Poisson sampling case by
0p zz {0* = (r,/5')' : r e R+,/3 £ Rp)i where the symbol R+ represents the
set of positive real numbers. It is important to note that &m ^0? since
constrains r to equal N while 0p requires r only to be positive.
The incomplete-data Poisson log likelihood can be written as
4P)(0*;y) = Y,{y^°svi^*) - vi(o*))
i
= X¡y¿(losr - 1°s(SI,íex pW)) + log(Lí exp(^))) -
it i
= y+ logr — t + XI y* los W exp(^)) - y+ los (S exp(^))
i i
= Mr) + 4(/3)>
(4.3.13)
where £2(/3) is defined to be the multinomial log likelihood in (4.3.12) and
£\(t) is the log likelihood for the Poisson random variable Y+ which is the
total sample size N. Since /3 is unconstrained for both sampling schemes,
we can find the ML estimates by differentiating (4.3.12) and (4.3.13) with
respect to /3 and finding the roots of these score functions. But the two score

- 157-
functions are identical implying that the maximum likelihood estimates of ¡3
are the same for both sampling schemes. That is, if we let /3(M) and /3(p)
denote the ML estimates of (3 under the multinomial and Poisson sampling
schemes, respectively, we have shown that /?(M) = ¡3(p\ Also, by (4.3.12) and
(4.3.13), we see that upon differentiating a second time
d2
df3'd(3
d2
M/3) -
d(3'd(3
so that the portion of the information matrix that pertains to /3 is the same
for both sampling schemes. Further, equation (4.3.13) shows that the log
likelihood for incomplete Poisson components can be expressed as a sum
of two, parameter independent, log likelihoods. Thus, the parameters are
orthogonal in that the information matrix is block diagonal, i.e. the parameter
estimates are asymptotically uncorrelated. The inverse of a block diagonal
matrix is simply the block diagonal matrix of the individual inverses. Hence,
the estimated variance of the ML estimates of (3 is the same for either sampling
scheme.
Cell Mean Inference. Notice that not only is M) = /3(p) but also
t(m) — f(p) = N. This follows since, in the multinomial case, r is necessarily
equal to the total sample size iV, while in the Poisson case, ^i(r) is simply
the log likelihood of the random variable Y+ which is Poisson with mean
r, implying that the ML estimate is f(p) = Y+ = N. However, we must
acknowledge the fact that the asymptotic variance of f under the Poisson
assumption is approximately N (it is var(Y"+), where Y+ ~ Poisson{r)),
while the variance of f under the multinomial assumption is zero (var(iV) = 0
since N is nonstochastic). This is important because inferences about cell

- 158 -
means (or cell probabilities) involve all of the loglinear parameters, even r.
Thus the variance of the cell mean estimates will depend upon which sampling
scheme is used.
Briefly, using the EM algorithm, we can find the observed information for
the loglinear parameters (a,¡3')' based on the assumption that the complete
data are product Poisson. The complete data means Hi are assumed to follow
the loglinear model
log Hi = a + x'i/3, i = 1,..., n.
If the sampling design is such that X+ = N, the total sample size, is fixed
so that X ~ Mult(iV, 7c(a,/3)) then the parameter a is ‘fixed by design’.
Actually, upon reparameterization, we see that /3 is free of constraints but
that a = a(/3, N), i.e. a is a function of /3 and N. In fact,
« = log(£Hi) ~ log(£ exp(x'/3))
i i
= logN- l°g(¿ exp(z¿/3))
(4.3.14)
Our objective is to find an estimate of the variance of the cell mean estimates (1
under the multinomial assumption. The calculation of this variance estimate
is complicated somewhat since the variance estimate of á is different for the
two sampling schemes. It is a simple application of the delta method to find
the variance of ¡1 under the Poisson assumption since ¡j, = exp(dln + X/3).
This follows since we’ve found the information for (a,/3) and hence the
estimated variance-covariance matrix of (a, /3) based on the assumption that
the complete data are product Poisson and that the incomplete data are of
the form Y = LX with L satisfying the same four properties as above.

- 159 -
Since, upon convergence of the EM algorithm, we compute the variance-
covariance matrix of (a,/3) under the product Poisson assumption only, we
must find a way to rewrite /i in terms of /3 and N only. But by (4.3.14) we
have the relationship
d = log N - l°g(¿ exP(x'ifi))
so that
A = exp(dln + Xf3)
= exp (lnlogiV + l„log( ——^-) +X¡3
\ £exp
= N( exp(X/3)^ \ =Ni exp(X/3) \
^E?exp(x^)/ \Vnexp(X¡3)J
(4.3.15)
Now since the information for (3 is the same under both sampling
schemes, we can find an estimate of the variance of ¡3 assuming the complete
data are multinomially distributed. We will actually find the variance of 7r,
which is nothing but -fo = (exp(X/3)/l'n exp(X/?)), via the delta method.
Delta Method. Since the ML estimate ¡3 is consistent, a first order
approximation to 7r can be found by using a Taylor’s expansion about the
true parameter value (30, viz.
7T « 7t(/50) + ~ Po)
Thus, the variance of if is approximately
var(-7r) « var^7r(/30) + ^i\p0CP ~ Pofj
A A
where var(/3) is that portion of the variance-covariance matrix of (d,/2)
pertaining to $. Recall that it was shown above that this portion is the
same for both sampling schemes.

- 160 -
It is shown in Appendix B that
= [Pi*) ~ â„¢']X (4.3.16)
where X = Z[,-lj. That is, X is the design matrix with the first column
deleted. Hence, the variance of ir under the multinomial assumption can be
estimated by
varMult(7r) = [D(tc) - ññ'](Xvax(j3)X')[D(Tt) - 7T7r'] (4.3.17).
4.4 Latent Class Model Fitting—An Application
To further illustrate the utility of the above results, we explore the fitting
of loglinear latent class models. For an expository on latent class analysis,
see Haberman (1979).
Suppose we can observe (manifest) factors Ai, A2,..., Ap with Jl512,..., Ip
levels, respectively, while a latent factor W with K levels is not observable.
Consider the set of cells, C = {(1,1,..., 1,1), (1,1,..., 1,2),..., (Jj,..., Ip)}
resulting from a cross classification on factors A\,..., Ap. Listing the elements
of C in lexicographical order, we denote the first cell by 1, the second by 2, and
so on to m, where m = II?=i U- With this representation the complete data
(the K *m cell counts) are X = (Xn,..., XiK,...,Xm,...,X!BK)T. The
observed data, Y, are the marginal counts collapsed over latent factor W.
Here Y = LX = (Xi+,..., Wm+)T, where L = 1TK ® Jm.
We initially assume that X is composed of independent Poissons with
means following the loglinear model,
logMj(ff)=a + x'ijP, » = l,...,m, j = l,...,K.

- 161 -
We can use the EM-algorithm of (4.3.7) to derive 9 = (a,/?')' and equation
(4.3.11) to obtain an estimate of its variance. From (4.3.8) the adjustment
matrix is Z'vax(X\LX — y)Z with
(Vi
0
0
... 0 \
\ax(X\LX - y) - =
0
V2
0
... 0
u
0
0
... Vm)
where
Vi =
wS(i-S)
y HK A*jl
Mi+ A*i+
P; + A»i+
W+
y* M.+ w+
-y-iSLiSE \
y* m¿+ /*¿+
^¡2 t*iK
~Vi
' Vi+ #*¡+
HiK f-, HiK \
Vi m+ C1 m¿+ ) /
MK
(4.4.1)
Notice that is the covariance of a Fí x 1 multinomial vector with index
y¿ = £i+ and cell probabilities {/Xy//2f+, j = 1,..., FQ.
Let 9 denote the final estimate of 9 obtained upon convergence of the
EM-algorithm. Using (4.3.11) and (4.4.1), we can derive an explicit estimate
of the variance-covariance matrix of 9. It is
(z'D(pi»))Z - Z'( ®a, 1, (4-4-2)
which is the inverse of the information matrix evaluated at 9.
Numerical Example. We consider the example introduced in section
4.1. The observed data are counts resulting from cross-classifying the 216
respondents with respect to whether they tend toward universalistic (1) or
particularistic (2) values in four different situations (A,B,C,D) of role conflict.
The data are displayed below in Table 4.1.

- 162 -
Table 4.1. Observed cross-classification of 216 respondents with respect to
whether they tend toward universalistic (1) or particularistic (2) values in
four situations (A,B,C,D) of role conflict
Observed
Observed
A
B
C
D
frequency
A
B
C
D
frequency
1
1
1
1
42
2
1
1
1
1
1
1
1
2
23
2
1
1
2
4
1
1
2
1
6
2
1
2
1
1
1
1
2
2
25
2
1
2
2
6
1
2
1
1
6
2
2
1
1
2
1
2
1
2
24
2
2
1
2
9
1
2
2
1
7
2
2
2
1
2
1
2
2
2
38
2
2
2
2
20
We illustrate the results of the previous sections by fitting a simple
loglinear latent class model to the data. The ordinary two-level latent class
model fitted by Goodman is equivalent to the loglinear model
logAbifcit — A1 + Af + + A k
\ n
where i,j, k, l, and t run from 1 to 2.
Using the notation defined above, the set of observable cells is
C = {(1,1,1,1),(1,1,1,2),(1,1,2,1),...,(2,2,2,1),(2,2,2,2)} and m = 24 =
16. The complete data are x = (xn, X12, • • •, £i6i, where for instance
X42 = £11222 represents the count in cell (1,1,2,2,2). Although, we assume
that the complete data means satisfy the model in (4.4.3), we are only able
to observe y = Lx where L = 1'2 ® Ji6. Hence, we will fit the model using

- 163 -
the EM algorithm defined in (4.3.7). The FORTRAN program em.loglin was
used to fit the model. The input information needed is
(1) m(0), an initial estimate of the complete data means
(2) m and n, the length of the observed and complete data vectors
(3) p, the number of independent loglinear parameters
(4) Z, the design matrix
(5) L, the mxn matrix that satisfies Lx = y.
As discussed in section 4.3.1, a simple initial estimate of p, and hence
of /3, is one that satisfies Lp^ = y. But, by simply allocating approximately
a half of each observed cell count to the two levels of the latent factor, we
can find a pW that satisfies Lp^ = y. This initial estimate of p also allows
us to omit the direct input of the observed data which can be obtained via
V°) = y.
The two-level latent class model fit the data well (G2 = 2.72, df= 6)
thereby giving us a simple way of interpreting the association among the four
situations of role conflict. Table 2 displays the model parameter estimates
and their estimated standard errors. To make model (4.4.3) identifiable, those
parameters not displayed in Table 4.2 were set to zero. The last column,
entitled “Unadj Std Error”, contains the standard error estimates that would
be used if the complete data were actually observed. These are too small and
are invalid.

- 164 -
Table 4.2. Parameter and Standard Error Estimates
Parameter
Estimate Std Error Unadj Std Error
A*
0.532
-0.911
A?
0.712
A?
0.604
A?
1.884
Af
3.160
\AW
A22
-4.032
\BW
A22
-3.444
\CW
A22
-3.126
\DW
A22
-3.081
0.491
0.276
0.197
0.177
0.225
0.171
0.212
0.168
0.334
0.237
0.530
0.317
3.593
1.543
1.151
0.563
0.962
0.518
0.603
0.386
Estimates of certain classification probabilities and their estimated
standard errors were also computed. These probabilities are defined as
=*-++++. = p(w=t)
= 7ri+++i/7r++++i = P(A = H W = t)
*u'W = 7T+1 ++,/*++++< = P(B = 1| W = t)
++1+t/*++++« = P(C = 1\W = t)
4'"' = 1r+++lt/1r++++, =P(D = 1\W = t)
The standard errors were found using the arguments of section 4.3.3 and the
delta method. For example, the conditional probabilities have form
b\7T
bin’’
where b\ and b2 are 1 x n vectors of known constants. Thus, by a direct
application of the delta method, an estimate of the asymptotic variance is
'b27tbi - b\irb2''1
var
/6l7t\ _ b27tbi - &J-7T&2
\&2*V L (£>27t)2
var
(tt)
(&2
(4.4.4)
where var(7r) is the variance of fr under the multinomial assumption, i.e.
equation (4.3.17). Actually, since the conditional probabilities do not involve

- 165 -
the intercept parameter, the variance of 7r under the Poisson assumption,
which is
772 var (£) = -^D(fj,)ZvaT(a,P)Z'D(v)
could be used in expression (4.4.4) and the result would be the same. This is
not true of the marginal probabilities which have form b^TV. An estimate of
the variance of 61 ir is
var(617r) = 61var(7r)6'1,
where var(7r) is the variance of 7r under the multinomial assumption. The
estimate would be inflated if one used the variance under the Poisson
assumption, reflecting the stochastic nature of the total sample size. To
illustrate, we consider an extreme example. Let b\ = l'n so that b\ñ = 1.0
with probability one. That is, i>i7r is nonstochastic. If we use the multinomial
variance estimator we get zero as our estimate of the variance. This is what
we know it to be. On the other hand, using the Poisson variance estimator
we get some positive value as our estimate of the variance. This is known
to be incorrect. The estimated probabilities and their estimated standard
deviations are displayed in Table 4.3.
Table 4.3. Classification Probability Estimates (Standard Errors)
Latent
Class t
*V
~A\W
7rl t
"it
KC\W
"it
*D\W
7rl t
1
.279 (.058)
.993 (.025)
.940 (.066)
.927 (.066)
.769 (.095)
2
.721 (.058)
.714 (.040)
.330 (.050)
.354 (.049)
.132 (.038)
From these estimated classification probabilities, we see that level 1 of
the latent class W can be labeled the ‘universalistic’ level. That is, subjects

- 166 -
in level 1 of the latent class tend to have universalistic views for all four
situations. Notice that, given a subject is in level 1 of the latent class, the
probability that they respond ‘universalistic’ is estimated to be at least .77
for each of the four situations. Similarly, one could label level 2 of the latent
class as the ‘particularistic’ level. Except for situation A, the estimated
probability that an individual in latent level 2 responds ‘particularistic’ to
the situations is at least .65. Since the latent class model (4.4.3) fits well,
we conclude that, given a person is intrinsically particularistic or intrinsically
universalistic, their responses to the four situations (A, B,C,D) of role conflict
are independent.
4.5 Modified EM/Newton-Raphson Algorithm
In this section we present an alternative root finding algorithm for
the incomplete exponential family score functions of equation (4.2.9). As
mentioned above, the EM algorithm has both positive and negative features.
Two very important positive features are (1) the EM algorithm is insensitive
to starting values and (2) the EM algorithm finds a root that maximizes
the likelihood. In contrast, since the incomplete-data log likelihood is not
generally a concave function of the parameters, the Newton-Raphson (NR)
or Fisher-scoring (FS) algorithms may not converge to a maximal root. In
fact, they will be very sensitive to starting values and may not converge at
all. Negative features of the EM algorithm include its slow convergence and
lack of precision estimate by-product. On the other hand, the NR and FS
algorithms work well locally, in that if we implement these methods very near

- 167 -
a maximal root, the convergence, relative to EM, is fast and an estimate of
precision of the ML estimator is obtained as a by-product.
In practice, the EM algorithm may quickly approach a small neigh¬
borhood around a maximal root, but then slowly converge to the root.
For this reason, we present an alternative algorithm that uses both EM
iterations and NR (or some modified NR, such as FS or quasi-NR) iterations.
Specifically, the EM algorithm will be used initially and then, upon reaching a
neighborhood of the maximal root, the NR type algorithms will be employed.
Meilijson (1989) suggested this approach in a fine expository of root finding
methods for incomplete data score equations.
Recall that when the complete data has distribution in the regular
exponential family the incomplete-data log likelihood has form (4.2.8), i.e.
hiPiy) = c*((3;y) - c(/3)
and that the score function has form
sY(P; y) = ¿M/3; v) = e„(t(x)\y = „)- e,(t(x)) (4.5.1)
To solve for a maximal root of (4.5.1) we can begin by using the EM
iterative scheme described in (4.2.11). We will conclude that the iterate
estimate is in a sufficiently small neighborhood of the maximal root as soon as
||/5M _£(m+i)|| < SWITCH(TOL), where SWITCH(TOL) > TOL of (4.2.1).
At this point, we will employ the iterative scheme described in (4.2.13). As
a first step in (4.2.13), we must calculate the matrix Ay(/3(m); y) which is an
estimate of the negative Hessian of the incomplete-data log likelihood. At
times the Hessian or expected Hessian can be explicitly calculated. This is

- 168 -
true in the Poisson loglinear case (see equations (4.3.9) and (4.3.10)). Thence
the matrix Ay(/?(m);y) can be explicitly calculated and inverted. Generally,
however, the matrix Ay will only be an approximation.
Since both Ep(T(X)\Y = y) and Ep(T(X)) must be calculated during
the EM algorithm, in view of equation (4.5.1), we must have the ability
to calculate Sy(/3;y) at different values of ¡3. We then could use as an
approximation to Iy(/9(m);y),
Sy(/?M+ei;
e«||
where the bracket notation B[i,] represents the ith row of matrix B and
= (0,..., 0, e, 0,..., 0)' is a p x 1 vector with a small number e in the ith
position. The value of e should be determined by rules used for numerical
differentiation. Meilijson (1989) discusses this approximation technique and
refers to it as EM-aided differentiation.
Evidently, if one uses approximation (4.5.2), the only functions needed
to be calculated for (4.2.13) are the score functions which are differences
between the conditional and marginal expected values of the sufficient statistic
T(X). Finally, upon convergence of (4.2.13) we can use [Ay(/?(°°h y)] 1 as an
estimate of the precision of the ML estimates /3.
If one feels the EM algorithm will converge quickly enough or that
the matrix inversion of Ay is unnecessarily burdensome, then one can
select SWITCH(TOL) = TOL. In which case, Ay will be inverted just
once, since the iterative scheme (4.2.13) will converge after one iteration.
For SWITCH(TOL) = TOL, the modified algorithm is simply the EM
algorithm supplemented by a single calculation of a precision estimate. If
y)
i = 1,
(4.5.2)
Ay(fi^y)[i,\ =

- 169 -
SWITCH(TOL) > TOL, then the EM algorithm can be viewed as a procedure
for finding an appropriate starting value for the faster iterative schemes such
as NR or FS.
The modified iterative scheme can be described as follows
(1) Solve for /?("+1) in E^+t)(T(X)) = E0(m)(T{X)\Y = y)
(2) If ||/3(m) - || > SWITCH(TOL), then replace /3(m) by /3(m+1)
and go to (1). Else go to (3).
(3) Calculate [Ay(/3(m); y)]_1 and Sy(/3(m);y) as discussed above. (4.5.3)
(4) Replace /3(m) by /3(m+1) = /3(m) + [Ay(/3(m); y)]_15y(/3(m); y)
(5) If ||/3(m) - /3(m+1)|| > TOL, then go to (3) (or (1))*. Else stop.
* If the faster, less stable, algorithms are having trouble converging, reset
SWITCH(TOL) to a smaller value and reuse the EM algorithm to get into a
smaller neighborhood of the maximal root.
Algorithm (4.5.3) should be stable, insensitive to starting values, rela¬
tively fast, and will provide an estimate of the precision of the ML estimate
as a by-product.
As a special case, let us consider applying the modified algorithm (4.5.3)
to the Poisson loglinear model of section 4.3. In that case we were able
to derive an explicit formula for the observed and expected information for
the incomplete data. For simplicity, we will use the expected information
displayed in equation (4.3.10) as our AY matrix, i.e.
Ay(0; V) = r)) = Z'D(M(fl)L'D-‘(LM(/3))LD(fi(/3))Z. (4.5.4)

- 170 -
By expression (4.3.5), we can write the score function as
SY(f}-,y) = Zy-L'(V^)], (4.5.5)
where the and ‘—’ are componentwise operators.
To start the algorithm, we apply the EM iterative scheme of (4.3.7),
continuing until ||/3(m) - /3(m+1)|| < SWITCH(TOL). At this point we will go
to step (3) of (4.5.3) using the formulas (4.5.4) and (4.5.5) for Ay and Sy.
Repeat steps (3)-(5) of (4.5.3) until the convergence criterion is met.
4.6 Discussion
This chapter emphasized loglinear model fitting when the data are
incomplete. As an example, a latent class loglinear model was fit to the
data presented in Goodman (1974). The primary method of obtaining
ML estimates of the loglinear parameters was the EM algorithm, but other
possibilities such as the Newton-Raphson algorithm were discussed.
In section 4.2 we reviewed the EM algorithm with special attention
given to the regular exponential family. For the regular exponential case, the
iterative scheme (4.2.11) was shown to be equivalent to the EM algorithm.
Then, in section 4.3.1, we derive the specific form for the EM algorithm
when the data are product Poisson with means following a loglinear model.
An explicit formula for the observed information matrix is derived in section
4.3.2. An estimate of the variance of the ML estimates of latent class loglinear
parameters is shown in equation (4.4.2).
The assumption that the data are product Poisson is not as restrictive
as it may seem. In section 4.3.3 we discuss inference for loglinear parameters

- 171 -
when the complete data are multinomially distributed. The results follow by
arguments of Birch (1963) and Palmgren (1981). It is shown that, when the
total sample size is considered fixed, inferences about all loglinear parameters,
except the one that is fixed by design, are the same for both the product
Poisson assumption and the multinomial assumption. A method of estimating
the variance of classification probability estimates (and functions thereof) is
also developed in this section.
We introduce an alternative root finding algorithm (4.5.3) for the
incomplete exponential family score functions in section 4.5. The algorithm
exploits the positive features of both the EM and Newton-Raphson type
algorithms. Specifically, the algorithm should prove to be insensitive to
starting values and relatively fast (compared to straight EM). It also will
provide an estimate of the precision of the estimators as a by-product.
As mentioned above, many models that can be fit using the EM
algorithm can also be fit more directly using the Newton-Raphson algorithm.
Appendix B includes a discussion about the program NLIN which fits
generalized linear-nonlinear models. Also included in the appendix, is
the code for the two model fitting programs ‘em.loglin’ and ‘NLIN’. The
FORTRAN program ‘em.loglin’ is based on the iterative scheme (4.3.6) and
the formula (4.3.9) for the observed information matrix. The Splus program
‘NLIN’ can be used to fit generalized linear and nonlinear models. The
data are required to be independent and of the exponential dispersion type
(see discussion of NLIN). The author plans on implementing the algorithm
described in (4.5.3) for the Poisson loglinear model case.

APPENDIX A
CALCULATIONS FOR CHAPTER 2
We set out to show that the matrix of equation (2.3.11), viz.
DM -agd
_iOi£j 0
is equal to the matrix
ÍD(,To
)-®7T0i7r{)i 0\ / -D(tto)
0 Oj l g'(ío) o
-l
( Mi 0 \
Vo M2)
where
M, = D-\7T0) - D-1(7To)ii(JH-'D-1(7ro)Lf)-1if'D-1(7ro) - ®f 1*1'
R
and
Proof: For notational convenience, let D = D{7r0) and let H = H(£0). We
will state a basic matrix algebra result, the proof of which can be found in
Aitchison and Silvey (1958).
Let A be nonsingular and B be of full column rank. Assuming
compatibility
(A -BY1 - [A-1 -A-'BiB'A-'By'B'A-1 A-1B(B,A~lB)-l\
\-B' 0 ) “V {B'A-'BYB'A-1 -(B'A-'B)-1 )m
That is, the partitioned matrix has a simple inverse.
- 172 -

- 173 -
Using this result, identifying D and H/n» with A and B, we arrive at
an equivalent form for (2.3.11). It is
(D-1 - D-'HiH'D-'Hy'H'D-1 n*D-lH(H'\
^ n^H'D-'Hy'H'D-1 -n\(H'D~'H)-1 ) X
( D - ©TTo^ 0\
V 0 oJx
(D-1 - D-'HtH'D-'Hy'H'D-1 n^D^H^H'D^H)-1 \
V n^H'D^H^H'D-1 -nl(H'D~l H)~{ )’
Now, using the fact that D_1(7r0)( © 7r0¿7ró¿)D-1(7ro) = ®1rVr =
(ffilpX©!#) and, by Lemma 2.3.1, (©1^)# = 0, we can multiply out these
three partitioned matrices to get
( Mi 0 \
VO M2)
where
Mi = D-\tt0) - D-'MHiH'D-'MHy'H'D-'M - ©f 1*1'*
and
4
M2 = nl{H,D-1(%0)H)-1.
This is what we set out to show.
Result 3 (2.4.6) We wish to show that the asymptotic variances are related
~(P)~(P)'
var(/¿(M)) = var(/i(p)) - ©f ————
Tli
according to

- 174-
Proof: Since /j, = e^, we can invoke the delta method to arrive at
var(/i(Afl) = var(e í(Af)) = .D(eí°)var(£(M))£>(eí°)
= .D(e^0) ^var(£(p)) - ffi—.D(e^0), by 2.4.5
Í i'
= .D(e^0)var(£(p)).D(e^0) - 0^°*£o*
ni
~(p)~(py
= var(A^)-©^
Tli
where the equal signs represent asymptotic equivalence.
Result 4 (2.4.7) We wish to show that the asymptotic variances of the
freedom parameter estimates are related according to
var(/3(M)) = var(/3(p)) - A,
where
A = (X'Xy'X'C
Q^p.)C'X(X'X)-1.
Vni
Proof: In the following, the equal signs represent asymptotic equivalence.
Now, since ¡3 = (X1 X)-1 X' C log^Afi), we can invoke the delta method to
arrive at
var(/?(M))
= (X'X)-1 X'Cv ar( log(AA(M)))C"X(X'X)-1
= (X'X)-1A:,CL>-1(A/io)Avar(A(M))A'D-1(A/ro)C"X(X'X)-1
= (X'X)-1X'CD-\AnQ)Av3,v(ii^)A'D-\An<))C,X(X'Xy1
K~(p)~(p)\
© ^ £ JA'D-yA^C'XiX'X)-1
= var(/^p^)
, ^(P)~(P)'
- {X'Xy'X'CD-yA^AÍ © ^ £ j A!D-1{Ah0)C,X{X,X)-\

- 175 -
But by assumption (Al) of section 2.3.1,
D-1 (An)A( ® ^l^A'D-1 {Ah)
- D-1 ( ( ®Ali \ ( J
\ ®A2jHj ) \®A2j) V ® Vñj
— JJ-! ( ©Ajj/Zj \
V ®A2jHj )
( ffiAim
Vñj ’ v^i
\ n-i f ®^1¿¿í¿
; V ®A2jPj
(
Vn;
lTi lm2
\®A/
Í ffi Ik?
V® vV
Hence, we have that the asymptotic equivalence
var(/3(M)) = var(/3(p)) - A
holds, where
A = {X'Xy'X'C
(0 lmi \
A-
m lm2
\®vs-/
(®^- ^)CT(M)"
which is what we set out to show.

APPENDIX B
CALCULATIONS FOR CHAPTER 4
We prove that the four properties of the EM algorithm introduced in
section 4.2.1 do indeed hold. These proofs are essentially those of Dempster
et al. (1977) and Little and Rubin (1986).
Property 1. If 6and 0(m+1) are the mth and m + 1st iterate estimates
obtained via the EM algorithm then
eY(S^-,v) > £Y{«(m)-,yy,
i.e. the log likelihood is increased at each successive iteration.
Proof: As in section 4.2.1, we write the incomplete data log likelihood as
eY(6;y) = Q(e,e(m);y)-H(e,e<."')-,y).
Now, by Jensen’s inequality, H(6,6^;y) < 0(m); y), V0. This follows
since
(0(m), »(m); v) = iW = J l°gfx]Ax\ 0(m>)fxir(x; 0(m,)A/
= / {log (y^ir) +l°zfxirW)}fx¡Y(*;»(m))d
= / l0S ( fxly(X: ^ ^ + HW"h > H(»,»(");»)
(B.l)
- 176 -

- 177-
where the last inequality holds since the ‘log’ function is concave whereby
Jensen’s inequality tells us that
/log ^Xfx\Ax-o) ~) 9{m)^du - ~ los / fx\Y(*; 6)du = loS 1 = °-
Now, equation (B.l) holds at 6 — 0(m+1), i.e.
tf(0(m+1),0(m);y) <
Therefore,
£y(0(m+1); y) = Q(6(m+1\ y) - y)
> Q(6(m+1\ ^(m); y) - if y)
> Q(0, y)
= iy(«(m,;y)
where the second inequality follows since 0(m+1) is defined to be that value of
9 that maximizes the function Q(#,0(m);y). Hence we have shown that
Property 2: The sequence of EM iterates >1} satisfy, whenever
y(m) converges to as m -> oo,
¿4(«;y)U., = Sr(«(“);í/) = o.
i.e. the estimates converge to a zero of the score vector for Y.
Proof: Using Property 3, we can write the score vector for the incomplete
data as
ír(«(”,);j/) = ¿Q(M(m);y)U..>.

- 178 -
But this implies that
^fr(M(m,;y)U->=o,
since
!') = gjM»;!/)l*-, = ^«(M("); y)U>
Therefore,
= 0 + o(l;m —► oo)
since, by definition of 0(m+1),
and because as - 0(m+1)|| goes to zero the function 0(m); y)|^m+1)
goes to zero. But by convergence properties of the EM algorithm ||0(m) -
0(™+i)|| -»0asm-too, Thus equation (B.2) holds and is tantamount to
§q¿y{9-,v) I*») =0.
Property 3: For any 60,
o,»)U = Sy(e 0; y) = E*,(sx(e« X)\Y = y).

179 -
Proof:
¿Q(M.;y)| = E9o(px(«-,X)\ei)lY = y)
= Ell'(Sx(e0;X)\Y = y)
= I Sx(0o•,x)fX\Y{x-,y,9Q)dv
J R
= IRWlosfx{x’e)
fx{x;0)
Jr fx{x) 0)dv
)
dv
_ jRWefx(x;0)dv
Jr fx{x'i Q)dv
&0
= á(Iog/,/x^;S)d,/)l».
= ¿l°g/y(y;^)l»„ = SY(S0;y).
Property 4: For any 0o,
Ir(«o; y) = £*(/*(«.; X)ir = y) - var»0 (SX(0O; X)\Y = y).
Proof: Since the observed information matrix is the negative Hessian of
the log likelihood, we have that
lr(0; y)
dHY(e-,y)
BO'dO
33 Q(M«;y)-g|^tf(Mo;y)
dB'de
E‘° (- wml*V'< x>|r=v)-E*(- wmeM»<». =y)
= £,„ (/*(»; X)\Y = y) - £fc(JX|K(«;y, X)|K = »).

- 180 -
But
Eao(ixiy(«a-iy,x)\Y = ») = £*(- = »)
= ^(^|y(#¡»-y)k^x|y(tf;w-y)Ulir = »)
= ^([Sx(9„;X)-Sy(«0;y)]x
[5x-(9o;X)-5y(90;y)]'|y = y)
= £*([Sx(»„;X) - E„t(Sx(e0-,X)]Y = y)]x
[Sjr(»„;.X) - Ee„(Sx(e„-X)\Y = y)]'\Y = y)
= B9„(M»o;^)^(9o;X)|y = y)
- £*,(S*(0„;X)|y = y)Eg0(S'x(6o] X)\Y = y)
= var„o(5A-(90;X)|y = y).
Hence
M»o! y) = E<,a(Ix(6-,X)\Y = y) - var^S* (90; X )|y = y),
which is what we set out to show. g
Theorem: If the complete data vector X has distribution in the regular
exponential family, i.e. the density function has form
fx(x\P) = a(x)exp(T'(x)P-c(J3)) (B.3)
with respect to some measure, then the EM algorithm can be used to find the
MLE of P based on incomplete data Y = Y(X) and the algorithm is as stated
in (4.2.11).
Proof: Sundberg (1976) shows that the EM algorithm can be used to find
the ML estimates of P based on incomplete data. We will show that the

- 181 -
general EM algorithm of (4.2.3) reduces to (4.2.11) when the complete data
have distribution in the regular exponential family.
The general EM algorithm (4.2.3) is defined as
Q(/3(m+1),/3(m);y) = max Q(0,pW ¡y)
where
Q(t3,/?<"•>;!/) = E0{m)(ix(p-,X)\Y = y).
Now since X has density of form (B.3), it follows the the log likelihood
ix{/3',X) has form
txtfi X) = loga(X) + T'(X)(3 - c(/3).
Hence,
Q(P,/3 Now, since
dp d0 dpdp vzr„U{X))
is negative definite, it follows that the solution, say /3(m+1\ to
^Q(P,0^;y) = 0
is the value of (3 that maximizes the function Q(/9,/3(m); y). But
^pQ(P,0(m);y) = E^(T(X)¡Y = y) -
= E„{m>(T(X)\Y = y) - E,(T(X)).

- 182 -
Hence /?(m+1) satisfies
Ep(m+i)(T(X)) = Ep(m){T(X)\Y = y)
which is tantamount to showing the equivalence of the two iterative schemes
(4.2.3) and (4.2.11). a
We differentiate the score vector of equation (4.3.5) to obtain an explicit
expression for the observed information matrix. Recall that we are to show
that the information matrix can be expressed as in (4.3.9), viz.
Ir(0;y) = Z'D(n)L'D(j±-)LD(»)Z - Z’D(L'(^ii))DMZ.
Proof: By equation (4.3.5), we know that the score vector for Y is
â„¢e(^)^
Now
dSY{(3;y) _ (dSY(py)\ ( dp\
dp V ap )\dp)'
where
d\i _ dexp(Z(3)
dp ~ W
= D(n)Z.
In the following, denote the n x 1 elementary vectors by e¿. That is,
e' = (0,0,.. .,0,1,0,. ..,0,0),
where the ‘1’ is in the ith position.

- 183 -
We set out to find the derivative of the score vector with respect to /j,.
It is
d-MM = JL[z'DM±{y^)L,]
dfi1
L\n
Therefore,
=9-W(py^^
= Z'(e’!e'i)[pyij^)LlSIn)
= Z'D(pyijg£)L,)
m
-xm'LWjfcpW
= Z'd(l\?-L»-
Ly,
-ZWD^)L.
9SY(/3-,y) _ fdSY((3;y)\ ( dfx\
d/3' V dyf )\dp)
= Z’D(n)L'D(-^)LD(n)Z - Z'd(l'(Z^¿))d(v>)Z,
which is what we set out to show.
Using the delta method we can find the asymptotic variance of 7r. The
expression for the asymptotic variance involves the matrix dir/d(3'. We show

- 184 -
that equation (4.3.16) holds, i.e.
dir _ Vn/j.)-nt¿]X _
d(3'
i1^)2
= [D(ir) - 7T7r']X.
Proof: From (4.3.15) we have that
„ = exp(al. + X/3) = N (),
or equivalently that
_/£ _ / exp(X/3) ^
"U.
iV Vl^exp(X/3)/'
Here ¡3 is an unconstrained parameter vector of length p. Notice that l'nfj, = N
and hence,
dir _ d / exp (X/3) \ _ d / ¡j, \
W ~ W\Vnexp{X(3)J ~ W\K¿)
= W(¿®^)+K(I^)(^(1W)
= ¿W-(¿FÍ‘1”W
= ¿ w - ^'x
= [JM(g;= \d(«)-™‘]x. m
DESCRIPTION OF COMPUTER PROGRAMS
em.loglin. Briefly, em.loglin is a FORTRAN program that can be used to
obtain ML estimates of loglinear parameters as well as an estimate of their
precision when only disjoint sums of the complete Poisson data are observable.
The EM algorithm (4.3.7) is used to find the ML estimates and expression
(4.3.9) is used to calculate the precision estimate. It is assumed that the

- 185 -
complete data X are distributed product Poisson with n x 1 mean vector
p following the loglinear model log p = Z/3. The incomplete data must be
expressible as Y = LX where L is an m x n matrix that satisfies properties
(l)-(3) of (4.3.1). The user must input the following information
(1) an initial estimate of the complete data means that equation
L/j(°) = y, i.e. /i(°) is consistent with the observed data y
(2) m and n, the length of the observed and complete data vectors.
(3) p, the number of loglinear parameters
(4) Z, the n x p full column rank design matrix
(5) L, the m x n matrix that satisfies Y = LX.
The output includes
(1) /3, an ML estimate of the loglinear parameter vector /3
(2) var(/3), an estimate of precision of the ML estimate
(3) G2, the likelihood ratio goodness-of-fit statistic
(4) df, the degrees of freedom associated with the null asymptotic
Chi-square distribution of G2
(5) p, an estimate of the complete data cell means
(6) var(/l), an estimate of the precision of p (Poisson sampling)
NLIN NLIN is an Splus (Becker, et al. 1989) program that fits generalized
linear and nonlinear models to data with distributions in the exponential
dispersion family (Jqrgenson, 1989). We now briefly describe exponential
dispersion models and how to fit them.

- 186 -
A General Algorithm For Fitting Generalized Linear-Nonlinear Models. Let
Yi,...,Yn ~ ind ED(m,a2,Wi),
i.e. the density function for the random variable Y¿ has form
f{Vi) = a(y¿,<72,w¿)exp{^(0¿y¿ - k(0¿))}>
where k(0) is the cumulant function and /c'(0¿)d=r(6i) = Hi-
Suppose that each mean can be expressed as an invertible function of
some covariate vector and a p x 1 parameter vector, i.e. Hi = /i¿(xj,/3), i =
1,..., n. Some examples are
(1) Hi = x'ift, Linear Model, Identity Link
(2) Hi — exp(x¿/3), Linear Model, Log Link
(3) Hi — exP(cc¿/3)/(l +exp(x¿/3)), Linear Model, Logit Link
(4) Hi — exp(X/3), Nonlinear Model, Log Link
Example (4) is nonlinear when the matrix L is some m x n (m < n) matrix
satisfying (4.3.1) and is not the identity matrix. Note that L\ is the ith row
of the matrix L. In fact, the matrix L can be chosen so that the Poisson
loglinear latent class models are a special case of example (4).
Letting the vector h = (hi,..., hn)' and the symbol ED represent a
particular exponential dispersion distribution, we say that {ED, h} specifies
a generalized linear-nonlinear model. As a special case, suppose that each \
has a common inverse g such that
K1(Hi) = gM, t = i,...,n.
We say that the triple
{ED,ri = g()i),rH = x',pi,
(BA)

- 187-
specifies a generalized linear model (GLM) (McCullagh and Nelder, 1989).
In GLM parlance, the function g is known as the ‘link’ function. Examples
include
(1) Poisson Loglinear Model:
{Poisson(/r), r¡ = log(/¿), = x'fi}
(2) Binomial Logistic Model:
{Binomial(n,7r), g = logjf^, = x'fl}
(3) Normal Linear Model:
{Normal^, £), rj = fi, rj{ = x'fi}
Maximizing the Likelihood Our objective is to make inference about the
loglinear parameters in (3 and hence about the means We will base our
inference on the maximum likelihood estimates and their precision. Therefore,
we must maximize the log likelihood with respect to /2. The log likelihood for
the sample Y is
= ¿loSa{yi, i i
where k'(9í) = /i¿ = h^x'^P).

- 188 -
The score function is
_dl{p-y) _ 1 "
s(P\y) =
d(3
1 V''«. \f \(„. \
i A ,dm^ i
where
d(3 )Kdni
= ¿ií )(»-*)
= ¿¡Ew‘(Tjp)v~tM«
=^w)wv"s
= XD'WV-'S,
S = y-fi
w = ®"wi
(B.5)
D =
d/¿
Here the matrix D is referred to as the ‘model matrix’. The maximum
likelihood estimate may be found by solving for a zero of the score function
(B.5) (at least in many cases). To solve for this zero, we will use a Newton-
Raphson type algorithm which will require calculation of the Hessian matrix.
W1=hU*™1»
=^wl%{{v~IWD) 8 h)+(s' * ^é{v~'WD)
= ±-D'(-V-1WD + Z)
a2
= Xi-D'V-'WD + D'Z)
a2
where E(Z) = 0 so that the expected value of the Hessian is
e( — —D'V~lWD
V dP'd(3 ) a2±JV VV±J-

- 189 -
Therefore, for /3(fc) in a neighborhood of ¡3, the solution to the score equation,
we have the following linear approximation
dWk+1)\y) „ di(/m-,y) 1 dH^y){p dp
_i_
cr2
dp ' dp'dp
D'WV-'S - D'WV-lD(p(k+r> - pW)
i=L{p<)
The next estimate of ¡3 will be p(k+1\ the solution to the linear equation
L(p(k+J)) = 0. The solution is
^(fc+i) _ ^(k) + {pwv-lD)-1i>wv-1s
= (D'WV~1D)~1D'WV~1(Dp(-k^ + S) (B. 6)
= {D,WV-lD)-lD,WV~li^
where = DP^ -|- S is a ‘local’ dependent variable.
The iterative algorithm (B.6), which is the Fisher-scoring algorithm, is
also referred to as the iteratively reweighted least squares algorithm (IRLS).
The reason for this label is evidently due to the last expression in (B.6).
For each k, it resembles a weighted least squares estimate, where the weight
matrix is W-1, the model matrix is D, and the dependent variable is
Denoting the ML estimate by ¡3, we have that in many situations
¡3 ~ AN{P,a\D'WV-1D)-1),
i.e. P has an asymptotic normal distribution. Also, we let consistent estimator of the dispersion parameter cr2. For example, dividing
the deviance statistic by the degrees of freedom associated with its asymptotic
distribution results in a consistent estimator of cr2 (Jprgenson, 1989).

- 190 -
By evaluating D and V at ¡3 and using the consistent estimator can consistently estimate the asymptotic variance of ¡3 by
var(/3) « o2(b'WV~lb)~l.
The astute reader will notice that, upon specification of the exponential
dispersion distribution, the matrix V is determined. Also, the matrix W
is a matrix of known constants. Hence, the only matrix not determined as
yet is D, the so called ‘model matrix’. The matrix D is a function of f3 and
X through the following function
D = w = wh(x'^
When the model is of the form (B.4), i.e. the model is a GLM, we have that
the model matrix
i- ( \(dM\
dp Kdg’WJK dp )
_ ( ( dr] \ _ (drj'Y1
-\dr,')\dp)-\dn)
and can be calculated explicitly. But, more generally, when the model is
{ED,h}, D can not be calculated explicitly or at least is very difficult to
calculate explicitly. However, it can be numerically estimated.
Numerical Approximation to D. We use a popular and simple technique to
numerically approximate D. Recall that D is the matrix of partial derivatives
of fi with respect to /3. Hence, the problem is to approximate a derivative
matrix. One such estimate, and the one used in the program NLIN, is
D « Dn = [n((3 + e1)-^i((3-e1),...,n((3 + ep)-^-ep)\E 1 (B.7)
where = (0,..., 0, e, 0,..., 0)' is a p x 1 vector with the small constant e in
the ith position and the matrix E is a p x p diagonal matrix with 2e on the
diagonal. Thus E = 2elp = 2[ei,..., ep\.

- 191 -
Now the IRLS algorithm will involve just one additional step and that is
to calculate a numerical approximation to the model matrix D. The actual
algorithm used in NLIN is
(1) Input y,w,fj, = h(X,(3),V(ix), and the deviance function Dev(y,w, y)
(2) Find an initial estimate /3(°) of ¡3
(3) Compute D= Dn(/3^), = V(/3(ml), and
— y - /i(/3(ml) (B.8)
(4) Compute /3(m+1) =
(5) Compute Dev(y,w,
(6) If ||Dev(y, w, - Dev(y,wy > TOL, replace ¡3
by and go to (3). Else stop.
Notice that step (1) of (B.8) involves inputting the data, the weights,
the mean function, the variance function, and the corresponding deviance
function. It follows that this program can more generally be used to fit models
via quasi-likelihood methods (McCullagh and Nelder, 1989). Another remark
is worthwhile mentioning. When the model is {ED, g(n) = /i, fj. = X/3}, i.e.
a Linear, Identity link model, the numerical approximation Djv of D in (B.7),
which equals X, is exact. Specifically, for the Normal Linear Model

- 192 -
the approximation is exactly equal to the model matrix X. The argument is
as follows
(DN)ij — [Vi(P + ej) ~ Vi{P - ei)]/2||ei||
Thus, Dn = X = D.
= WÁP + Cj) - x'AP - *j)\/2e
= [AP + ¿¿i - AP + Aej]/2e
= 2x\ey/2e = x\ejle
~ exij/e=xij
= (X)y = {D)a

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1049-1060.

BIOGRAPHICAL SKETCH
Joseph Benedict Lang was born in St. Cloud, Minnesota, on February-
12, 1963. In 1967, his parents, Ralph and Mary Jean Lang, moved the family
to Richmond, a small resort town in central Minnesota. He remained in the
central Minnesota area for 23 years. His parents, 7 sisters, and 1 brother
remain there to this day. In 1982, he decided to pursue a college degree. His
10 year “career” as bartender and cook looked to be nearing an end when
he began his post-secondary education at St. Cloud State University. After
a brief period of entertaining the idea of majoring in art, Joseph grew very
fond of mathematics and statistics and decided to focus his attention on these
more quantitative disciplines.
After receiving his Bachelor of Arts degree in mathematics from St.
Cloud State University in 1986, Joseph was encouraged to pursue his Master’s
and Ph.D. degrees in statistics at the University of Florida in Gainesville.
He went on to receive a Master of Statistics degree in 1988 and, under the
direction of Professor Alan Agresti, was awarded a Ph.D. degree in statistics
in the spring of 1992. While working toward these degrees, he worked as
a teaching assistant, biostatistics consultant, and a research assistant. In
1992, Joseph accepted an academic position as assistant professor in the
Department of Statistics and Actuarial Science at the University of Iowa.
- 200 -

I certify that I have read this study and that in my opinion it conforms to
acceptable standards of scholarly presentation and is fully adequate, in scope
and quality, as a dissertation for the degree of Doctor of Philosophy.
(
(vJ
(r
Alan i
vgresti, Chair^
Professor of Statistics
I certify that I have read this study and that in my opinion it conforms to
acceptable standards of scholarly presentation and is fully adequate, in scope
and quality, as a dissertation for the degree of Doctor of Philosophy.
Jétne Pendergast
Associate Professor of Statistics
I certify that I have read this study and that in my opinion it conforms to
acceptable standards of scholarly presentation and is fully adequate, in scope
and quality, as a dissertation for the degree of Doctor of Philosophy.
Rocco Ballerini
Associate Professor of Statistics

I certify that. I have read this study and that in my opinion it conforms to
acceptable standards of scholarly presentation and is fully adequate, in scope
and quality, as a dissertation for the degree of Doctor of Philosophy.
Carole Kimberlin
Associate Professor of Pharmacy
Health Care Administration
This dissertation was submitted to the Graduate Faculty of the
Department of Statistics in the College of Liberal Arts and Sciences and
to the Graduate School and was accepted as partial fulfillment of the
requirements for the degree of Doctor of Philosophy.
May, 1992
Dean, Graduate School

UNIVERSITY OF FLORIDA
3 1262 08554 0424



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