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TH E918

This is to certify that the
thesis entitled

PARENTAL DIVORCE AND DISORDERED EATING: AN
INVESTIGATION OF A GENE-ENVIRONMENT
INTERACTION

presented by

JESSICA LYNN SUISMAN

has been accepted towards fulﬁllment
of the requirements for the

Master of Arts degree in Psychology

 

 

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5/08 K:IProj/Acc&Pres/CIRCIDateDue.indd

PARENTAL DIVORCE AND DISORDERED EATING: AN INVESTIGATION OF A
GENE-ENVIRONMENT INTERACTION

By

Jessica Lynn Suisman

A THESIS

Submitted to
Michigan State University
in partial fulﬁllment of the requirement
for the degree of

MASTER OF ARTS
Psychology

2010

 

 

ABSTRACT

PARENTAL DIVORCE AND DISORDERED EATING: AN INVESTIGATION OF A
GENE-ENVIRONMENT INTERACTION

By

Jessica Lynn Suisman
Objective: Previous research demonstrates an association between parental divorce and
disordered eating that has traditionally been conceptualized as environmental in nature.
However, it is possible that a gene-environment interaction may drive this relationship,
whereby parental divorce serves as an environmental “trigger” for pre-existing genetic
susceptibilities to disordered eating. This study examined this possibility by investigating
whether the heritability of disordered eating is signiﬁcantly higher in children living in
divorced versus intact families. Methods: Participants included 1,810 adolescent and
adult twins (mean age = 18.26) from the Michigan State University Twin Registry and
the Minnesota Twin Family Study. Disordered eating was measured with the Minnesota
Eating Behavior Subscale (MEBS). Univariate, twin constraint models were used to
compare the heritability of disordered eating in divorced versus intact families. Results:
A gene-environment interaction was not observed for most measures of disordered
eating. However, a gene-environment interaction was suggested for body dissatisfaction,
as the heritability of body dissatisfaction was higher in twins of divorced versus intact
families. Discussion: Although gene-environment interactions do not appear to be
important for relationships between divorce and most forms of disordered eating, the
presence of these effects for body dissatisfaction requires further study. Future
investigations should replicate our results and begin to identify the factors underlying the

unique gene-environment relationships between divorce and body dissatisfaction.

TABLE OF CONTENTS

LIST OF TABLES .................................................................................... iv
INTRODUCTION .................................................................................... 1
METHOD .............................................................................................. 5
RESULTS ............................................................................................ l 3
DISCUSSION ....................................................................................... 16
REFERENCES ...................................................................................... 35

iii

 

LIST OF TABLES

Table 1 ................................................................................................ 24
Table 2 ................................................................................................ 25
Table 3 ................................................................................................ 26
Table 4 ................................................................................................ 27
Table 5 ................................................................................................ 30
Table 6 ................................................................................................ 31

iv

Parental Divorce and Disordered Eating: An Investigation of a Gene-Environment
Interaction
Introduction

Research in the past several decades has examined the psychological effects of
parental divorce on offspring (e.g., depression; Amato, 2001; Amato & Sobolewski,
2001; Hetherington, Bridges, & Insabella, 1998; Hetherington, Stanley-Hagan, &
Anderson, 1989). One area of focus has been on the association between eating disorders
and parental divorce (Billingham & Abrahams, 1998; Boumann & Yates, 1994; Herzog,
1982; Igoin-Apfelbaum, 1985; Martinez-Gonzalez, et al., 2003; Yannakoulia, et al.,
2008). For example, several studies have found higher rates of divorce in individuals with
bulimia nervosa as compared to controls (Boumann & Yates, 1994; Herzog, 1982; Igoin-
Apfelbaum, 1985), and that divorce prospectively predicts increased risk for onset of an
eating disorder (e.g., eating disorder not otherwise speciﬁed, bulimia nervosa and
anorexia nervosa; Martinez-Gonzalez, et al., 2003). Associations between divorce and
disordered eating extend to community samples as well, as body dissatisfaction and binge
eating are signiﬁcantly associated with parental divorce (Billingham & Abrahams, 1998;
Yannakoulia, et al., 2008).

Taken together, the literature suggests that there are cross-sectional and
prospective relationships between parental divorce and disordered eating that warrant
additional investigation. Studies are needed to examine aspects of the relationship
between divorce and disordered eating that would enhance understanding of divorce’s
effects. Researchers have not previously examined etiologic factors underlying

associations between parental divorce and disordered eating. In general, divorce is

 

viewed as a stressful life event that results in the accumulation of many negative events,
such as moving, decreases in socioeconomic status, changes in schools, changing
relationships with parents, decreased social support, exposure to parental conﬂict, and
loss of contact with extended family (Amato, 1993; Hetherington, 1993). Consistent with
these ideas, it has been hypothesized that the effects of parental divorce on disordered
eating are environmental in origin and lead to stress, increased negative affect, and
dysphoria, which may increase risk for disordered eating (Martinez-Gonzalez, et al.,
2003; Welch, Doll, & Fairburn, 1997).

However, these associations could also be due to gene-environment interactions,
where divorce serves as an environmental “trigger” for disordered eating in individuals
who have existing genetic susceptibilities. Individuals who do not have these genetic
predispositions, on the other hand, may be less likely to develop disordered eating in
response to a parental divorce. The signiﬁcant heritability of eating disorders (> 50% for
Anorexia Nervosa and Bulimia Nervosa; Bulik, Sullivan, & Kendler, 1998; Bulik, et al.,
2006; Bulik, Sullivan, Wade, & Kendler, 2000; Kendler, et al., 1991; Kendler, et al.,
1995) and disordered eating (e.g., binge eating, body dissatisfaction, weight
preoccupation; Bulik, et al., 1998; Culbert, Burt, McGue, Iacono, & Klump, 2009;
Klump, Burt, McGue, & Iacono, 2007; Klump, Suisman, Burt, McGue, & Iacono, 2009;
Sullivan, Bulik, & Kendler, 1998; Wade, Wilkinson, & Ben-Tovim, 2003) provides
partial support for this hypothesis. Moderate-to-high heritability estimates support the
possibility of a gene-environment interaction because they suggest some genetic
inﬂuence on the trait in question, which is necessary for gene-environment interactions to

occur. Despite calls from previous researchers to examine gene-environment interactions

for eating disorders (Bulik, 2005; see Bulik, et al., 2000; Klump, Wonderlich, Lehoux,
Lilenfeld, & Bulik, 2002; Wade, Bulik, & Kendler, 2001), no studies to date have
examined gene-environment interactions for disordered eating and divorce.

Parental divorce may interact with genetic risk as either a shared (i.e., co-twins
within a twin pair have the same experiences and this increases sibling similarity) or
nonshared (i.e., co-twins within a twin pair have different experiences that decrease
sibling similarity) environmental risk factor. The objective experience of parental
divorce is shared by co-twins within a twin pair, as both twins experience divorce within
their family. However, parental divorce could also be described as an effective nonshared
environmental factor, if the reaction to the divorce itself differs between siblings in a twin
pair and results in sibling differences in behavior, mood, etc. (e.g., one twin experiences
signiﬁcant distress following the divorce while the other experiences very little distress;
Turkheimer & Waldron, 2000). In the classical twin study in which gene-environment
interactions are not speciﬁcally examined, gene-shared environment interaction effects
load onto the heritability estimate (Purcell, 2002), which can lead to inﬂated heritability
estimates and decreased estimates of shared environment. When the risk environment is
instead nonshared in nature, the gene-nonshared environment interaction loads onto the
estimate of the nonshared environment (Purcell, 2002).

Given that divorce may be operating as either an objective shared or an effective
nonshared environmental factor, this particular interaction could be “hiding” in the
heritability or nonshared environmental estimates in previous twin research. As noted
above, heritability estimates have been moderate—to-high for disordered eating (see Bulik,

et al., 2000), and interestingly, nonshared environmental factors have been found to

account for the remaining variance. These ﬁndings collectively suggest that gene-
environment (either shared or nonshared) interactions may be present and could be
contributing to one or both of these signiﬁcant estimates. The present study will directly
examine this possibility by investigating whether gene-environment interactions between
divorce and disordered eating exist, and if so, whether divorce seems to be operating
primarily as a shared or nonshared environmental factor.

Previous research examining disordered eating and divorce has also not examined
the age of offspring at the time of the divorce, despite literature suggesting that there
tends to be differences in response to divorce in children versus adolescents
(Hetherington, et al., 1989). Given that the peak age of onset of disordered eating is
adolescence and early adulthood (American Psychiatric Association, 2000), it seems
likely that parental divorce during these periods would be more strongly related to
disordered eating than parental divorce during early childhood, since there is closer
proximity between the divorce itself and the typical age of onset of disordered eating.
This hypothesis is supported by evidence that, though there are long term impacts of
divorce, an especially risky period seems to occur within the ﬁrst two years following
divorce. This increased period of risk seems to be driven by the sheer number of stressful
events occurring during this time (e. g., changing living arrangements, decreased contact
with one parent, attempts to understand the divorce, etc. Kelly & Emery, 2003).

Given the above, the present study examined the possible presence of a gene-
environment interaction for symptoms of disordered eating and divorce in 1,810
adolescent and young adult female twins. The presence of a gene-environment interaction

was examined by comparing all divorced to all intact families. It was hypothesized that a

signiﬁcant gene-environment interaction would exist, where the heritability of disordered
eating symptoms would be higher in twins who experienced parental divorce than those
who did not. Exploratory, secondary analyses were also conducted to examine the effects
of twins’ age at the time of divorce. For these secondary analyses, it was hypothesized
that the strongest gene-environment interaction effects would be observed in twins who
experienced divorce during adolescence/early adulthood (i.e., age 13 or greater) versus
chﬂdhood.
Method

Participants

This study used archival data drawn from two population based twin studies, the
Michigan State University Twin Registry (MSUTR; Klump & Burt, 2006), and the
Minnesota Twin Family Study (MTFS; Iacono, Carlson, Taylor, Elkins, & McGue, 1999;
Iacono, McGue, & Krueger, 2006). Sample characteristics from each of these studies are
described in Table 1.

Recruitment procedures for the MSUTR are detailed elsewhere (Klump & Burt,
2006), and therefore will only brieﬂy be described here. The MSUTR recruited
adolescent and young adult twins (ages 10-28) using ﬂyers/paid advertisements (25%),
recruitment mailings through the MSU Ofﬁce of the Registrar (27%) and recruitment
mailings using birth records (48%) through the Michigan Department of Community
Health (MDCH). Although most participants completed all study procedures in the
laboratory (95%), some subjects who were not able to travel to the lab participated by
completing a mailed packet of questionnaires. Importantly, participants from the MSUTR

have been shown to be representative of the population from which they were drawn in

terms of racial and ethnic background (i.e., 83% Caucasian; Culbert, Breedlove, Burt, &
Klump, 2008; Klump & Burt, 2006).

Previous research demonstrates differences in the heritability of eating disorder
symptoms in pre-pubertal versus adolescent twins and adults (Culbert, et al., 2009;
Klump, Burt, et al., 2007; Klump, McGue, & Iacono, 2003; Klump, Perkins, Burt,
McGue, & Iacono, 2007) Therefore, MSUTR twins under the age of 14 were excluded
from the present study, and twins between the ages of 14-15 years were included gnlLif
they were in mid-puberty or beyond at the time of study participation. Mid-puberty was
indicated by a score 2 2.5 (see Culbert, et al., 2009; Klump, et al., 2003) on the Pubertal
Development Scale (PDS; Petersen, Crockett, Richards, & Boxer, 1988). The PDS asks
participants to report on the extent to which physical markers e. g., body hair growth,
breast changes, onset of menarche) of puberty have occurred. The PDS exhibits good
psychometric properties (Petersen, et al., 1988), and the PDS total score correlates highly
(r = 61-67) with physician ratings of pubertal development (Petersen, et al., 1988).

The second source of data comes from the MTFS, a population based,
longitudinal twin study of same-sex twins and their parents (Iacono, etal., 1999; Iacono,
et al., 2006). The MTFS data used in the present study includes 1,456 twins at
approximately age 17. At the time of recruitment for the study, researchers identiﬁed
twins who were either 11 or 17 years old using Minnesota birth certiﬁcates. Recruitment
efforts resulted in the recruitment of 91% of twins who met age criteria for the study.
Some twin families were later excluded because they I) lived further than one day’s drive
from the MTF S lab or 2) the twins had been diagnosed with a mental or physical

handicap that would prevent them from participating in the day long laboratory visits.

Like the MSUTR sample of twins, the MTF S sample is representative of the population
from which they were drawn in terms of racial and ethnic background (i.e., 98%
Caucasian; Iacono, et al., 1999; Iacono, et al., 2006), and is also comparable to Minnesota
census data across multiple demographic domains (e.g., urban/rumal split, parent age,
ethnicity, and marital status; Holdcraft & Iacono, 2004). Further details on recruitment
methods for this study are available elsewhere (Iacono, et al., 1999; Iacono, et al., 2006).

For the current study, data from both the 11 year-old and 17 year-old MTF S
cohorts were used. Age 17 assessments were used for both cohorts, which corresponds to
the second follow-up assessment for 11 year-old twins and the intake assessment for the
17 year-old twin cohort. Including data from MTF S participants at age 17 is
advantageous for several reasons. It maximizes the sample size of post-pubertal twins in
the study, which is essential given the differential heritability of disordered eating in pre-
versus post-pubertal twins (Culbert, et al., 2009; Klump, Burt, et al., 2007; Klump, et al.,
2003). It also allows for the examination of twins during peak periods of risk for eating
disorders (American Psychiatric Association, 2000) and closely matches the average age
of the MSUTR sample of twins (see Table 1).
Measures

Zygosity determination. Zygosity determination methods differed slightly for
the MSUTR and the MTF S. Both samples used the Physical Similarity Questionnaire
(Lykken, Bouchard, McGue, & Tellegen, 1990; Peeters, Van Gestel, Vlietinck, Derom, &
Derom, 1998), which correlates 95-99% with zygosity measured via genotyping. In
addition to this zygosity questionnaire, the MTFS also used a staff opinion (based on

physical similarity of face shape, ear shape, hair color, and eye color), and an algorithm

based on measurements of cephalic index (i.e., ratio of head width to length), ﬁngerprint
ridge counts, and the ponderal index (i.e., a measure of leanness calculated as height in
inches/”weight in pounds) to determine zygosity. When the three MTF S measures were
not in agreement, a serological sample was taken to determine correct zygosity.
Importantly, cases of mistaken zygosity in either the MSUTR or MTFS would decrease
estimates of both heritability and gene-environment interactions, rather than increase
estimates. This is because mistaken zygosity leads to the inclusion of some DZ twins
within the MZ twin category, and some MZ twins in the DZ twin category. These errors
would decrease overall differences between MZ and DZ twin correlations, thereby
driving down estimates of genetic effects (given that genetic effects are implicated when
MZ twins are more similar than DZ twins on a given trait). These decreases in genetic
effects would then, by deﬁnition, decrease the gene-environment interaction as well
(since a trait must be heritable for the interaction to occur).

Disordered eating symptoms. Disordered eating in both samples was measured
using the Minnesota Eating Behavior Survey (MEBS; Klump, McGue, & Iacono, 2000;
von Ranson, Klump, Iacono, & McGue, 2005). The MEBS is a 30-item, self-report,
true/false questionnaire. It was designed for use with children as young as 10 years, and
has been shown to have excellent reliability and validity in adolescent and young adult
females (see below). The MEBS includes a total score (i.e., overall measure of disordered
eating) that is comprised of four subscales: body dissatisfaction (i.e., dissatisfaction with
body weight/shape), weight preoccupation (i.e., preoccupation with weight and dieting),
binge eating (i.e., thoughts of and/or engaging in binge eating), and compensatory

behaviors (i.e., the use of inappropriate compensatory behaviors in order to change body

weight/shape). The present study did not include the compensatory behaviors subscale
due to low item endorsement and low internal consistency reliability in younger subjects
(Klump, et al., 2000).

Previous studies have demonstrated good internal consistency for the total score,
body dissatisfaction, and weight preoccupation subscales in children and adolescents
(alphas =.78-.89). Binge eating demonstrates a slightly lower alpha, (65-69) that is still
within the acceptable range (von Ranson, et al., 2005). Studies also demonstrate
satisfactory concurrent validity, as scores from the MEBS and Eating Disorder
Examination-Questionnaire (EDE-Q) correlate moderately to highly on subscales that
measure similar constructs (correlation = .83 in 14 year old girls; von Ranson, et al.,
2005). Finally, girls with eating disorders generally report signiﬁcantly higher scores than
girls without eating disorders on all of the MEBS subscales (von Ranson, et al., 2005).

Parental marriage history. History of twin exposure to biological parental
divorce was measured via twin self report in both studies. Twins were asked whether
their parents were divorced and, if so, their age at the time of the divorce.

Data Analysis

Primary analyses: Independent sample t-tests were used to examine whether the
current data replicate previous studies demonstrating increased rates of disordered eating
in women who experienced parental divorce (Billingham & Abrahams, 1998; Boumann
& Yates, 1994; Herzog, 1982; Igoin-Apfelbaum, 1985; Martinez-Gonzalez, et al., 2003;
Yannakoulia, et al., 2008). These tests examined whether there were mean differences in

MEBS scores in twins from divorced versus intact families.

The possibility of a moderating effect of parental divorce on the heritability of
disordered eating was then examined using twin intraclass correlations and biometric
model-ﬁtting. Twin intraclass correlations were used to examine initial indications of
differences in genetic and environmental effects in twins from divorced versus intact
families. Genetic inﬂuences are implied if the monozygotic (MZ) twin correlations are
signiﬁcantly greater than the dizygotic (DZ) twin correlations. Shared environmental
inﬂuence is suggested when the MZ and DZ twin correlations are approximately equal
and are also signiﬁcant. Finally, nonshared environmental inﬂuence (which also includes
measurement error) is inferred when the MZ correlation is less than 1.00, and/or both the
MZ and DZ twin correlations are small and non-signiﬁcant.

Univariate, twin constraint models were then used to examine the relative
inﬂuence of additive, genetic effects (A), shared environmental effects (C), and nonshared
environmental effects (E) both within the divorced and intact groups independently of
one another, as well as differences between the divorced and intact groups. Models were
ﬁt to raw data using Mx (Neale, 1995). Using raw data allows for the inclusion of all twin
pairs, as it treats missing data as random (Little & Rubin, 1987). This is an advantage
over the use of covariance matrices (where pairwise deletion occurs for missing data), as
twin pairs can still be included in analyses even if one twin is missing data.

For the primary analyses, I initially estimated variances, means, and covariances
of the raw data to obtain a baseline estimate of ﬁt (-21nL) for each subscale of the MEBS.
I then ﬁt fully unconstrained (i.e., A, C, and E are allowed to vary across the divorced
and intact groups) and fully constrained (i.e., A, C, and E are constrained to be equal

across the divorced and intact groups) biometric models to the data to examine possible

10

group differences in genetic and environmental effects. The ﬁt of these biometric models
was compared to that of the baseline model (i.e., the -21nL of the baseline model was
subtracted from the -2lnL of the biometric models), resulting in a likelihood-ratio chi-

square test of goodness of ﬁt for the model. This chi square was used to calculate
Akaike’s information criterion (AIC; xz-de), a measure of model ﬁt versus model

parsimony for the constrained and unconstrained models separately. The fully
unconstrained model provided estimates of the relative inﬂuence of A, C, and E within
each group (i.e., without taking into account the effects in the other group). In order to
examine differences between the two groups, the relative ﬁt of the unconstrained and
constrained models were compared using AIC (i.e., the smallest AIC indicated the best
ﬁtting model) and an additional likelihood-ratio chi-square test of goodness of ﬁt. This
second chi-square test compared the fully unconstrained model to the constrained model
by subtracting the -21nL of the fully unconstrained model from the -2lnlL of the
constrained model(s).

Importantly, these model ﬁt comparisons allowed for the determination of the
presence versus absence of moderating effects of divorce. For example, if the fully
unconstrained models provided a better ﬁt to the data, it would suggest that there are
differences in the inﬂuences of A, C, and/or E across divorced and intact groups. By
contrast, if the fully constrained model provided the best ﬁt to the data, then there would
be no evidence for genetic moderation of divorce on disordered eating, as it would
suggest that A,C, and E do not vary across groups.

Notably, before conducting the model-ﬁtting analyses described above, I ﬁrst

examined potential differences in genetic and environmental inﬂuences on MEBS scores

11

between the MTF S and MSUTR samples. A fully unconstrained (i.e., A, C, and B were
allowed to vary across the MTFS and MSUTR samples) and a fully constrained (i.e., A,
C, and E were constrained to be equal across the MTFS and MSUTR samples) model
were ﬁt to the data. The fully constrained model provided a good ﬁt to the data for all
subscales, suggesting minimal sample differences in genetic or environmental effects.
Given this high degree of similarity, the samples were combined in all subsequent
analyses (data not shown).

Secondary analyses: Given previous research and theory suggesting that there are
differences in responses to divorce in children versus adolescents (Hetherington, et al.,
1989), secondary analyses were also conducted to examine the potential inﬂuence of twin
age at the time of divorce. First, independent samples t-tests were run to examine mean
level differences in disordered eating between twins who experienced divorce in
childhood versus adolescence. Second, potential differences in the inﬂuences of A, C,
and/or B were examined across three groups: childhood divorce (i.e., divorce at age 12 or
under), adolescent divorce (i.e., divorce at age 13 and up), and intact families. Twin
correlations and constraint models were used as described above to examine whether the
heritability of disordered eating is moderated by age at divorce. Three additional
“partially” constrained models were also computed for these analyses, where two groups
were constrained while the other group was allowed to vary. Speciﬁcally, childhood and
adolescence were constrained while allowing intact families to vary, childhood and intact
families were constrained while allowing divorced families to vary, and adolescent and
divorced families were constrained while allowing intact families to vary. These

additional constraint models indicated whether subgroups of divorced families could be

12

constrained to be equal to each other while still varying from the intact families (e.g., Can
childhood and adolescent divorce be constrained to be equal while still varying from
intact families?)

Results

Prior to all analyses, the MEBS body dissatisfaction and binge eating scales were
transformed (loglo x + 1) due to the positive skew of the data. Age was then regressed out
of all MEBS scores given the wide age range of our sample and research suggesting that
mean levels of disordered eating varies by age, with lower rates at younger ages and an
increase across adolescence (American Psychiatric Association, 2000; Jones, Bennett,
Olmsted, Lawson, & Rodin, 2001; Klump, et al., 2000).

Independent samples t-tests were conducted to examine whether there were mean
differences in MEBS scores between the MTFS and MSUTR. Results suggested that
there were no signiﬁcant differences in mean MEBS scores across studies for all
subscales, providing further support for combining the two samples. Independent samples
t-tests were also used to examine whether mean levels of disordered eating differed
between intact and divorced families. Results indicated differences at the trend level for
the total score and weight preoccupation, such that scores appeared to be higher in
divorced compared to intact families. Mean level differences did not emerge for body
dissatisfaction and binge eating (see Table 2), although the mean differences were in the
expected direction (i.e., higher in divorced group). Importantly, the lack of strong
phenotypic associations between divorce and disordered eating does not preclude the
possibility of etiologic moderation, as gene-environment interactions may attenuate

phenotypic associations.

13

Primary Analyses. As expected, twin correlations suggested genetic inﬂuences on
disordered eating regardless of divorce status, as the MZ twin correlations were, in
general, signiﬁcantly greater than the DZ twin correlations (see Table 3). Signiﬁcant
nonshared environmental inﬂuences were also implied, as evidenced by MZ twin
correlations that were less than 1.0. Finally, the shared environment appeared to be
negligible for all MEBS subscales, as the MZ twin correlations were generally at least
double the DZ twin correlations (i.e., the MZ-DZ correlations are not equal). Twin
correlations for the total score, weight preoccupation, and binge eating subscales were
quite similar between intact and divorced families (i.e., the difference between MZ and
DZ twin correlations did not appear to differ greatly across family type). Therefore a
gene-environment interaction is unlikely for these subscales. However, twin correlations
for the body dissatisfaction subscale suggested that a gene-environment interaction may
be present, as the difference between MZ and DZ twin correlations was greater in
divorced compared to intact families, suggesting increased heritability in divorced
families.

Results of model ﬁtting analyses are presented in Table 4. Notably, C was
estimated at 0 in all of the models (data not shown). These results are consistent with
previous research suggesting that the inﬂuence of C on disordered eating is negligible in
pubertal and post-pubertal twins (Bulik, et al., 1998; Culbert, et al., 2009; Javaras, et al.,
2008; Kendler, et al., 1991; Klump, Perkins, et al., 2007; Reichborn-Kjennerud, et al.,
2003). Therefore, models were re-run with C constrained to zero (i.e., estimating only A

and B). These models uniformly provided the best ﬁt to the data (lower AICs and non-

14

signiﬁcant chi-square difference tests) as compared to the ACE models. Thus, only the
AE models are presented in Table 4.

Overall, model ﬁtting analyses indicated that there were no signiﬁcant differences
in heritability in divorced versus intact families. The fully constrained AE models
provided the best ﬁt to the data for the total score, weight preoccupation, and binge eating
subscales, as indicated by the lower AIC value and the non-signiﬁcant change in chi-
square. Thus, while both genetic and nonshared environmental inﬂuences were important
for these types of disordered eating symptoms, there were no gene-environment
interactions that differentially inﬂuence genetic or environmental inﬂuences in the two
groups.

The one exception to this general rule was body dissatisfaction. The fully
unconstrained AE model provided a better ﬁt to the data than the fully constrained model,
as indicated by the lower AIC and signiﬁcant change in chi-square. This ﬁnding suggests
that A and/or E signiﬁcantly differed between the divorced and intact groups. In order to
further elucidate the nature of the effects, sub-models were ﬁt to the data to examine
whether group differences in A and B were statistically signiﬁcant. These models
estimated A while constraining E to be equal across groups and vice-versa. The ﬁt of
these models was then compared to the fully unconstrained model to determine the ﬁnal,
best ﬁtting model. Neither A nor E could be constrained across group, as evidenced by
the signiﬁcant change in chi-square and increased AIC as compared to the fully
unconstrained model. Therefore, a moderating inﬂuence of divorce on the genetic a_nd
nonshared environmental inﬂuences on disordered eating (i.e., a gene-environment

interaction) appears to be present for body dissatisfaction. Speciﬁcally, parameter

15

estimates indicated that genetic effects are greater in the divorced group (a2 = .76) than in
the intact group (a2 = .56). Conversely, effects of the nonshared environment were greater
in the intact group (e2 = .44) than in the divorced group (e2 = .24).

Secondary Analyses: Age of Divorce. Findings for age of divorce were
remarkably similar to those described above. T-tests revealed no signiﬁcant differences in
mean levels of symptoms of disordered eating by age at divorce (Table 5), and twin
correlations suggested that timing of divorce is unlikely to have an impact on heritability
of disordered eating symptoms (Table 3). The pattern and magnitude of M2 and DZ
correlations for the childhood and adolescent groups were similar (see Table 3). Model
ﬁtting analyses are presented in Table 6. These analyses support indications from the
twin correlations, as the fully constrained model (i.e., intact, childhood divorce, and intact
divorce groups constrained to be equal) was the best ﬁtting for all scales except for body
dissatisfaction. For body dissatisfaction, the model constraining childhood and
adolescent age at divorce to be equal provided the best ﬁt, while the model constraining
intact families to be equal as well did not provide a good ﬁt. These results suggest that a
gene-environment interaction was present for body dissatisfaction only, and age at
divorce did not impact these ﬁndings.

Discussion

This was the ﬁrst study to examine gene-environment interaction effects of
parental divorce on the heritability of disordered eating. Evidence for a gene-environment
interaction did not emerge for most symptoms of disordered eating (i.e., total score, binge
eating, and weight preoccupation). Only body dissatisfaction exhibited these effects, in

that that the heritability of body dissatisfaction was higher in offspring of divorced than

16

intact families. Age at the time of parental divorce did not impact any ﬁndings. Taken
together, the present study Suggests that the experience of divorce is associated with
increased heritability of body dissatisfaction, but not other forms of disordered eating,
regardless of age at divorce.

At the phenotypic level, results suggested only modest associations between
disordered eating and divorce. No signiﬁcant mean differences across divorce status were
observed for binge eating and body dissatisfaction, and only trend-level differences were
present for total score and weight preoccupation. These ﬁndings suggest that mean level
effects, if present, are likely quite small. Indeed, effect sizes for all mean differences
ranged from .05-.11 (see Table 2). Small effect sizes may explain inconsistencies in
previous research, where some studies found associations between divorce and
disordered eating (Billingham & Abrahams, 1998; Boumann & Yates, 1994; Herzog,
1982; Martinez-Gonzalez, et al., 2003; Yannakoulia, et al., 2008) while others did not
(Dolan, Lieberman, & Evans, 1990; Johnson & Flach, 1985; Mitchell, Hatsukami, Eckert,
& Pyle, 1985). Large sample sizes would be needed to detect these small phenotypic
effects, and thus, some studies with smaller samples (Dolan, et al., 1990; Johnson &

F lach, 1985) may have failed to identify signiﬁcant associations.

Gene-environment interaction effects were only observed for body dissatisfaction
in this study. Results suggested signiﬁcantly greater heritability of this measure of
disordered eating in the divorced group. It will be important for future studies to both
replicate our results and examine mechanisms underlying these effects. For example, one
body of literature suggests that body dissatisfaction is linked to depression, and generally

suggests that individuals who are depressed have increased levels of body dissatisfaction

l7

(Allgood-Merten, Lewinsohn, & Hops, 1990; M. Cooper & Hunt, 1998; P. J. Cooper &
Fairburn, 1993; Joiner, Schmidt, & Singh, 1994; Joiner, Sclunidt, & Wonderlich, 1997;
Joiner, Wonderlich, Metalsky, & Schmidt, 1995; Keel, Fulkerson, & Leon, 1997; Keel,
Mitchell, Davis, & Crow, 2001; Leon, Fulkerson, Perry, & Cudeck, 1993; McCabe &
Marwit, 1993; Rierdan & Koff, 1997; Roth & Armstrong, 1993; Taylor & Cooper, 1986).
Associations between body dissatisfaction and depression appear to be unique, as there
are links between depression and body dissatisfaction even in the absence of other
symptoms of disordered eating (e. g., Keel, et al., 2001). This may help explain why body
dissatisfaction, but not other forms of disordered eating, showed unique gene-
environment interaction effects. Importantly, depression is also associated with parental
divorce (Hetherington, et al., 1998; Hetherington, et al., 1989; Huurre, Junkkari, & Aro,
2006; Starksen, Raysamb, Moum, & Tambs, 2005; Strohschein, 2005), and there is some
evidence that separation events in childhood (including parental divorce) exhibit gene-
environment interaction effects for depression (i.e., childhood separation events increase
risk for depression only if high latent genetic risk is present; Zimmermann, et al., 2008).
Given the above, it may be that a gene-environment interaction emerges for body
dissatisfaction due to a particularly robust association between body dissatisfaction and
depression, and interactions between depression and divorce. Unfortunately, this
hypothesis could not be directly tested in the present study, as different measures of
depression were used across twin registries (e.g., depression symptom counts in MTFS;
Beck Depression Inventory (BDI) in MSUTR) and age groups (BDI in adult MSUTR
twins; Children’s Depression Inventory (CD1) in adolescent MSUTR twins). Future

studies should directly investigate this hypothesis.

18

The gene-environment interaction effects observed for body dissatisfaction did
not emerge for any other measures of disordered eating (i.e., MEBS total score, binge
eating, and weight preoccupation). It appears that the modest phenotypic effects of
divorce on disordered eating do not translate into etiological differences between groups
for these measures, despite theories proposing gene-environment interaction effects
(Bulik, 2005; Klump, et al., 2002). Similar results have emerged for other psychological
variables, whereby environmental rather than genetic causal mechanisms have been
linked to divorce, even given signiﬁcant heritability of the phenotype (e.g., delinquency;
Burt, Barnes, McGue, & Iacono, 2008; D'Onofrio, et al., 2007; D'Onofrio, et al., 2005,
2006; O'Connor, Caspi, DeFries, & Plomin, 2000). Future studies should measure
divorce in combination with other life stressors (e. g., decreased family income, changes
in family composition) to determine whether gene-environment interaction effects
emerge for these more comprehensive measures of divorce “stress”. Composite measures
of life stress have demonstrated gene-environment interactions for other disorders (e.g.,
Caspi, et al., 2003; Silberg, Rutter, Neale, & Eaves, 2001), and should be explored in
future research of disordered eating.

Given the lack of gene-environment effects for most disordered eating symptoms,
it will be important for ﬁsture research to consider explanations for these null effects.
One possibility is that the link between divorce and disordered eating is driven entirely by
enviromnental mechanisms. For instance, given that divorce is associated with numerous
stressful life events (e.g., moving, changing relationships with peers, decrease in
socioeconomic status, etc), and that stressful life events are linked to disordered eating

(Ball & Lee, 2000; Welch, et al., 1997), it is possible that the stress of divorce directly

19

increases levels of disordered eating. (Jacobi, Hayward, De Zwaan, Kraemer, & Agras,
2004). If the main effects of the environment are driving the association, environmental
risk factors are likely to be nonshared in nature, as the shared environment is generally
non-signiﬁcant for disordered eating in adolescence and adulthood (Klump, et al., 2009)
and the lack of signiﬁcant gene-environment interaction suggests that shared
environmental effects are not “hiding” in the additive genetic estimates. An alternative
possibility is that the association is driven entirely through the main effects of genetic risk
factors. Indeed, classical twin studies have suggested that the experience of divorce is
heritable (McGue & Lykken, 1992). The heritability of divorce is unlikely to operate
through a set of divorce “genes” per se, but may instead be transmitted through heritable
psychological traits (e.g., personality characteristics) that increase risk for divorce
(Jockin, McGue, & Lykken, 1996). With regard to disordered eating, it is possible that
the same heritable personality characteristics that increase risk for divorce may also
increase risk for disordered eating (e.g., high levels of negative affect). If this were the
case, main effects of genetic risk factors would be operating, as the same genes would
increase risk for divorce in the parents and disordered eating in offspring.

One ﬁnal possibility is that an association between divorce and disordered eating
is driven by gene-environment correlations rather than gene-environment interactions.
This hypothesis could not be tested in the present study, as statistical models do not exist
to test for gene-environment correlations when the moderator (i.e., divorce) does not vary
between twin pairs (Purcell, 2002). However, one particular type of gene-environment
correlation, passive gene-environment correlations, may be at play in this case. Passive

gene-environment correlations occur when offspring passively receive genes from their

20

parents that are associated with the environment in which they are raised. In the case of
disordered eating, it is possible that children receive genes from their parents that
increase their vulnerability to disordered eating while, at the same time, increases the
likelihood of exposure to divorce. Speciﬁcally, it is possible that a parent high in levels of
disordered eating may be more likely to experience marital dissatisfaction and divorce
(Friedman, Dixon, Brownell, Whisman, & Wilﬂey, 1999; Van den Broucke &
Vandereycken, 1989) than a parent low on disordered eating. In this case, the offspring
would be exposed to both the environmental (i.e., parental divorce) and genetic (i.e.,
genes for disordered eating) risk for disordered eating, and the two correlated risk factors
together may increase the likelihood of disordered eating in the offspring.

Future studies are needed to directly examine whether any of the effects discussed
above are operating, i.e., whether the main effects of the nonshared environment, the
main effects of genetic factors, or passive gene-environment correlations contribute to
disordered eating and divorce associations. Studies using other designs, such as adoption
studies, will be especially useful in examining which of these effects may be important.
For instance, an adoption study of divorced parents and adopted children could give some
indication of the importance of all three mechanisms. If the divorce of adoptive parents
still inﬂuences disordered eating of adoptive children, genes and passive gene-
environment correlations must not be contributing to the association, as adoptive children
and their parents by deﬁnition do not share genetic material. In this case, the main effects
of the environment would be paramount, as adoptive parental divorce could only have

environmental effects on their adoptive children’s disordered eating risk.

21

In addition to examining the overall effects of divorce on disordered eating risk,
the present study also was the ﬁrst to investigate whether timing of divorce is important.
Age at divorce failed to have an impact on mean levels of disordered eating symptoms or
gene-environment interactions between divorce and disordered eating. These ﬁndings
were somewhat surprising, given robust associations between adolescence and the
development of disordered eating (American Psychological Association, 2000).
However, these results may have been inﬂuenced by the timing of the assessment of
disordered eating. Disordered eating was assessed some time after divorce in this study
(M = 10.25 years post-divorce, SD = 5.23). It is possible that timing of divorce may
actually be important, but the effects are only detectable when measured directly after the
divorce, rather than many years later. Future research should investigate whether the lack
of effect found in the present study persists when disordered eating is examined during
the time period directly following the divorce.

There are some limitations of the present study that should be noted. Parental
divorce, but not marital discord, was examined. Parental divorce and marital discord are
strongly related (see Amato & Sobolewski, 2001), but they are often conceptualized and
studied separately, as divorce results in changes in family composition and other stressors
that are not present for marital discord alone (Amato & Sobolewski, 2001). Studies
including measures of both marital discord and divorce suggest that they have similar
effects on various outcome variables (e.g., psychological well being, relationships with
mother and father; Amato & Sobolewski, 2001), although this has never been examined
for eating disorders. Thus, it is unknown whether results of the present study would

differ if marital discord, rather than divorce, were examined for its moderating effects.

22

 

Future studies should include measures of parental divorce and marital discord in order to
elucidate the relationship between these two variables and commonalities and differences
in their association with disordered eating symptoms.

This study examined disordered eating using a non-clinical sample of subjects.
Given that subjects were not clinically diagnosed with an eating disorder, it is unknown if
present ﬁndings will generalize to clinical populations. However, estimates of genetic
and environmental effects from non-clinical samples are nearly identical to those from
clinical samples (Bulik, et al., 2000), suggesting the present results would likely be
similar in clinical populations. Further, it would be difﬁcult to directly examine
moderating effects of divorce on disordered eating symptoms in a clinical sample, as all
subjects would have high levels of disordered eating (e. g., body dissatisfaction, binge
eating, etc.), reducing variability in the outcome variables. However, future studies could
investigate the clinical signiﬁcance of these ﬁndings by investigating whether the
interaction of divorce and body dissatisfaction is predictive of the later development of
clinical eating disorders.

In summary, this was the ﬁrst study to directly examine gene-environment
interaction effects of parental divorce on disordered eating. Future research is needed to
clarify the magnitude and clinical signiﬁcance of phenotypic effects of divorce on risk for
disordered eating, given the small effect sizes detected in this sample. Studies using large
samples should also replicate gene-environment interaction effects for body
dissatisfaction and investigate potential mechanisms that drive the interaction (e. g.,
depression). Further, these ﬁndings should be extended to measures of marital discord

and other symptoms of disordered eating.

23

Table l.

MTFS, MSUTR, and Combined Sample Characteristics

 

MTF S vs. MSUTR

 

 

 

 

Combined Mean Comparisons
Sample MTFS MSUTR t(df) p.
N 1,810 1,456 354 -- --
MZ Pairs 1,125 925 200 -- --
DZ Pairs 685 531 154 -- --
Divorced Families: N 397 (22%) 303 (21%) 94 (27%) __ __
(%)
Intact Families: N (%) 1,413(78%) 1,153(79%) 260 (73%) -- --
Current Age
Range 14-28 16-20 14-28 -- --
Mean 18.26 17.85 19.94 12.15 <.001
(SD) (1.76) (0.70) (3.20) (358.33)
Age at Divorce
Range 1-22 1-18 1-22 -- --
Mean (SD) 7.92(5.13) 7.68 (4.92) 8.74 (5.75) 1.70 (382) .09
MEBS Total Score
Range 0-29 0-27 0-29 -- --
Mean (SD) 7.72 (6.06) 7.62 (6.04) 8.11 (6.12) 0.69 (1735) .49
MEBS Body
Dissatisggcmg
Range 0-6 0-6 0-6 -- --
Mean (SD) 2.40 (2.20) 2.35 (2.20) 2.58 (2.22) .29 (1735) .78
MEBS Weight
Preoccupation
Range 0-8 0-8 0-8 -- --
Mean (SD) 2.93 (2.48) 2.93 (2.50) 2.93 (2.39) .28 (1734) .86
MEBS Binge Eating
Range 0-7 0-7 0-7 -- --
Mean (SD) 1.34 (1.55) 1.30 (1.51) 1.52 (1.72) 1.53 (1734) .13

 

Note. MTF S = Minnesota Twin Family Study; MSUTR = Michigan State University Twin
Registry; MEBS = Minnesota Eating Behavior Survey; MZ = Monozygotic Twins; DZ =

Dizygotic Twins. N = Number of individuals included in sample. Though raw data are

presented for descriptive purposes, log transformed scores for weight preoccupation and
binge eating were used for T-tests in order to account for positive skew. Age was regressed
out of all MEBS variables prior to T-tests.

24

Table 2.
Mean Differences in Levels of Disordered Eating in Intact versus Divorced Families

 

 

Intact MEBS Divorced MEBS t Cohen’s

MEBS Scale Mean Mean (df) p d

(SD) (SD)

7.57 8.20 1.78 .10
Total Score (6.01) (6.16) (1721) .08
Weight 2.86 3.14 1.85 07 .11
Preoccupation (2.44) (2.59) (559.27) '
Body 2.35 2.57 1.33 19 .10
Dissatisfaction (2.17) (2.29) (1721) '

. . 1.32 1.39 1.05 .05

Binge Eating (1.55) (1.53) (1720) .29

 

Note. MEBS = Minnesota Eating Behavior Survey. Age was regressed out of all scores prior
to analyses, and log transformed scores were used for the binge eating and body
dissatisfaction subscales. Raw means and standard deviations are presented here for
interpretive purposes.

25

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29

Table 5.

Mean Differences in Levels of Disordered Eating Symptoms in Childhood Divorce versus

Adolescent Divorce Families

 

 

. . Adolescent
MEBS Scale ChlldhOOd Divorce Divorce Cohen’s
Mean (SD) NC”)
(N = 259-260) Mean (SD) d
(N = 89)
8.10 8.22 .09 (347) .93 .02
Total Score (6.12) (6.30)
Weight 3.06 3.19 .44 (347) .66 .05
Preoccupation (2.54) (2.70)
Body 2.58 2.45 .37 (347) .71 .06
Dissatisfaction (2.32) (2.22)
. . 1.38 1.39 .15 (346) .88 .006
Binge Eating (1.51) (1.60)

 

Note. MEBS = Minnesota Eating Behavior Survey. Age was regressed out of all scores prior

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30

 

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