Gender differences in the factor structure of posttraumatic stress disorder symptoms in war-exposed adolescents
DSM-IV's three-factor model of posttraumatic stress disorder (PTSD) is rarely empirically supported, whereas other four-factor models (King et al., 1998; Simms, Watson, & Doebbeling, 2002) have proven to be better representations of PTSD's latent structure. To date, a clear consensus as to which model provides the best representation of PTSD's underlying dimensions has yet to be reached. The current study investigated whether gender is associated with factor structure differences using the King et al. (1998) model of reexperiencing, avoidance, numbing, and hyperarousal PTSD symptoms. Participants were war-exposed Bosnian secondary/high school boys and girls (N=1572) assessed nearly two years after the 1992-1995 Bosnian conflict. Confirmatory factor analytic tests of measurement invariance across PTSD model parameters revealed many significant sex-linked differences. Implications regarding the potential role of gender as a moderator of the King et al. (1998) model's factor structure are discussed.
Journal of Anxiety Disorders 25 (2011) 604–611
Contents lists available at ScienceDirect
Journal of Anxiety Disorders
Gender differences in the factor structure of posttraumatic stress disorder
symptoms in war-exposed adolescents
, Jon D. Elhai
, Christopher M. Layne
, Mark Shevlin
, Elvira Durakovi
, Robert S. Pynoos
National Centre of Psychotraumatology, University of Southern Denmark, Odense, Denmark
School of Psychology, Faculty of Life and Health Sciences, University of Ulster at Magee Campus, Londonderry, Northern Ireland BT48 7JL, United Kingdom
Department of Psychology, University of Toledo, Toledo, OH, USA
UCLA National Center for Child Traumatic Stress, Department of Psychiatry and Biobehavioral Sciences, University of California, Los Angeles, USA
University of Sarajevo, Sarajevo, Bosnia and Herzegovina
Received 7 June 2010
Received in revised form 28 January 2011
Accepted 28 January 2011
DSM-IV’s three-factor model of posttraumatic stress disorder (PTSD) is rarely empirically supported,
whereas other four-factor models (King et al., 1998; Simms, Watson, & Doebbeling, 2002) have proven
to be better representations of PTSD’s latent structure. To date, a clear consensus as to which model
provides the best representation of PTSD’s underlying dimensions has yet to be reached. The current study
investigated whether gender is associated with factor structure differences using the King et al. (1998)
model of reexperiencing, avoidance, numbing, and hyperarousal PTSD symptoms. Participants were war-
exposed Bosnian secondary/high school boys and girls (N = 1572) assessed nearly two years after the
1992–1995 Bosnian conﬂict. Conﬁrmatory factor analytic tests of measurement invariance across PTSD
model parameters revealed many signiﬁcant sex-linked differences. Implications regarding the potential
role of gender as a moderator of the King et al. (1998) model’s factor structure are discussed.
© 2011 Elsevier Ltd. All rights reserved.
Posttraumatic stress disorder (PTSD) was ﬁrst introduced into
the Diagnostic and Statistical Manual (DSM) nomenclature in 1980
(American Psychological Association [APA], 1980). Since its initial
inclusion, the diagnosis of PTSD has been visited with controversy
(Spitzer, First, & Wakeﬁeld, 2007), including concerns relating to its
latent factor structure. Namely, although PTSD is currently repre-
sented in the DSM-IV (APA, 1994) as a constellation of 17 symptoms
grouped into three symptom clusters, this three-factor model has
rarely received empirical support (cf. Shevlin, McBride, Armour, &
Adamson, 2009). Indeed, the factor analytic literature to date has
overwhelmingly supported a four-factor structure for PTSD (King,
Leskin, King, & Weathers, 1998; Simms et al., 2002). Recent empiri-
cal studies addressing dimensionality of the construct of PTSD have
focused increasingly on evaluating whether the goodness of ﬁt of
PTSD models is contingent on key moderating variables. In particu-
lar, gender is strongly related to the likelihood of a PTSD diagnosis
(Tolin & Foa, 2006). However, the inﬂuence of gender as a potential
Corresponding author at: National Centre of Psychotraumatology, University of
Southern Denmark, Odense, Denmark. Tel.: +44 7876142643.
E-mail addresses: firstname.lastname@example.org, email@example.com
moderator of PTSD’s factor structure has yet to be systematically
The underlying dimensions of PTSD are currently represented
in the DSM-IV by three factors: Intrusion (Criteria B1–B5), Effort-
ful Avoidance/Emotional Numbing (C1–C7), and Arousal (D1–D5).
Empirical support for this tripartite latent structure is rare (cf.
Asmundson, Stapleton, & Taylor, 2004). In contrast, the majority of
factor analytic studies have provided support for four-factor mod-
els. The two models receiving the most attention and support to
date include that proposed by King et al. (1998), and that proposed
by Simms et al. (2002). The major differences between the two
models lie in their comparative placement of items D1 (sleeping
difﬁculty), D2 (irritability or anger), and D3 (concentration difﬁcul-
The King et al. (1998) 17-item PTSD model structure was derived
by allocating ﬁve items to an Intrusion factor (B1–B5), two items to
an Avoidance factor (C1–C2), four items to an Emotional Numbing
factor (C3–C7), and ﬁve items to an Arousal factor (D1–D5). This
model differs from the three-factor DSM-IV model in that it splits
the avoidance/emotional numbing factor into two separate fac-
tors based on evidence that emotional numbing and avoidance are
separate pathology-related constructs (Asmundson et al., 2004).
In contrast, the Simms et al. (2002) model groups the 17 PTSD
items by allocating ﬁve items to an Intrusion factor (B1–B5), two
items to an Avoidance factor (C1–C2), eight items to a Dyspho-
0887-6185/$ – see front matter © 2011 Elsevier Ltd. All rights reserved.
C. Armour et al. / Journal of Anxiety Disorders 25 (2011) 604–611 605
ria factor (C3–C7 and D1–D3), and two items to an Arousal factor
(D4–D5). This conceptualization is based on studies linking the Dys-
phoria factor items to other mood and anxiety disorders (Watson,
2005). Reﬂecting this similarity in content across disorders, recent
studies have found that the Dysphoria factor is less speciﬁc to
PTSD than the other three PTSD factors (Armour, McBride, Shevlin,
& Adamson, in press; Armour & Shevlin, 2010; Elklit, Armour, &
Shevlin, 2010). For example, when controlling for depression across
all PTSD items, Armour et al. (in press) and Elklit et al. (2010) both
found a signiﬁcant decrease in PTSD factor loadings, with the great-
est degree of attenuation occurring in the Dysphoria factor.
The current literature is ambiguous with respect to which of
the two models yields a more accurate representation of the latent
structure of PTSD. Speciﬁcally, ample support can be found for both
the King et al. (1998) model (most recently in Elhai, Engdahl, et al.,
2009; Lancaster, Melka, & Rodriguez, 2009; Naifeh, Elhai, Kashdan,
& Grubaugh, 2008; Palmieri, Marshall, & Schell, 2007; Schinka,
Brown, Borenstein, & Mortimer, 2007), as well as the Simms et al.
(2002) model (most recently in Armour & Shevlin, 2010; Boelen,
van den Hout, & van den Bout, 2008; Elhai, Ford, Ruggiero, & Frueh,
2009; Elklit & Shevlin, 2007; Palmieri, Weathers, Difede, & King,
2007). Accordingly, a general consensus as to which of these two
models provides the best representation of the underlying dimen-
sions of PTSD has yet to be reached.
As part of continuing efforts to clarify the strengths and poten-
tial drawbacks of the two models, recent studies have highlighted
a number of moderating variables that may be linked to the latent
structure of PTSD. Identifying potential moderators is of particular
value because it helps to clarify the circumstances under which a
given model ﬁts best, and by extension, to identify speciﬁc popula-
tions, settings, or applications for which the model may be better
suited. For example, studies have found that PTSD model ﬁt varies
as a function of measurement format (e.g., interview vs. self-report)
(Palmieri, Weathers, et al., 2007), and whether respondents are
instructed to rate PTSD symptoms based on their worst trauma vs.
their global trauma history (Elhai, Engdahl, et al., 2009).
In addition to these efforts to evaluate the comparative ﬁt of the
two PTSD factor models, a related line of studies have focused on
evaluating the stability (as gauged by the variance vs. invariance of
selected model parameters) of key model features as a function of
speciﬁc sample or setting characteristics. For example, the facto-
rial invariance of models of PTSD has been evaluated across groups
that differ in their native language (Marshall, 2004; Norris, Perilla,
& Murphy, 2001), military deployment status (Mansﬁeld, Williams,
Hourani, & Babeu, 2010; Simms et al., 2002), and era of military
service combined with treatment-seeking status (McDonald et al.,
2008). Of particular note, few studies have conducted compre-
hensive measurement invariance testing, but instead have tested
for model invariance in one or two focal model parameters (e.g.
McDonald et al., 2008).
To date, gender remains an understudied potential modera-
tor of the factor structure of PTSD. The relevance of gender to
the ongoing evaluation of competing models of PTSD is under-
scored by ﬁndings that although men tend to experience higher
rates of exposure to potentially traumatic events, women have
a two-fold higher risk of experiencing PTSD following exposure
to traumatic events (Breslau, Davis, Andreski, & Peterson, 1991;
Kessler, Sonnega, Bromet, Hughes, & Nelson, 1995). This two-fold
risk for PTSD in trauma-exposed women was supported by a meta-
analysis of 52 studies (Tolin & Foa, 2006). Women are also more
likely to develop chronic forms of PTSD compared to men (Breslau
& Davis, 1992; Breslau et al., 1998).
Few compelling explanations for these gender differences in
PTSD prevalence rates have been offered. One explanation centers
on the joint propositions that (1) different types of traumatic events
and circumstances differ in their etiological potency for developing
PTSD (cf. Layne et al., 2010); (2) men and women have differen-
tial rates of exposure to different types of trauma; and therefore,
(3) the speciﬁc types of trauma to which women have compar-
atively higher rates of exposure are more likely to lead to PTSD.
However, multiple studies have documented higher rates of PTSD
in women even after controlling for trauma categories that occur
more frequently in women, including sexual assault and domestic
violence (Breslau, Chilcoat, Kessler, & Davis, 1999; Fullerton et al.,
2001; Stein, Walker, & Forde, 2000).
A second explanation for differential prevalence rates in PTSD
between the genders is based on ﬁndings that women are twice
more likely than men to report depression and anxiety symptoms
(Tolin & Breslau, 2007). Accordingly, higher prevalence rates of
PTSD in women may reﬂect a higher general prevalence of psycho-
logical distress or of psychiatric disorder. However, Breslau, Davis,
Andreski, Peterson, & Schultz (1997) reported that gender differ-
ences in PTSD rates remained even after controlling for pre-existing
psychiatric disorders such as depression and anxiety.
A third explanation for gender-linked differences in PTSD preva-
lence rates centers on PTSD Criterion A, which, serves a key
“gatekeeper” function to a PTSD diagnosis. Speciﬁcally, Criterion
A is comprised of both objective (A1: traumatic experience) and
subjective (A2: intense emotional reaction) components. Given evi-
dence that women are more likely than men to disclose initial
adverse emotional reactions to trauma exposure by endorsing Cri-
terion A2 (Brewin, Andrews, & Rose, 2000), Criterion A2 may permit
gender to play a pivotal role in determining who qualiﬁes as having
experienced a Criterion A event.
In light of these gender-linked differences in variables linked to
the diagnosis or prevalence of PTSD, the current study examined
links between gender and the factorial invariance of an empiri-
cally supported model of PTSD. Of particular interest, both the King
et al. (1998) and Simms et al. (2002) models were originally devel-
oped using samples consisting of mostly or entirely adult men –
speciﬁcally military veterans. Notwithstanding this methodolog-
ical artifact, factor analytic support has been found in all-female
samples for both the King et al. (1998) model (Palmieri & Fitzgerald,
2005), as well as the Simms et al. (2002) model (Krause, Kaltman,
Goodman, & Dutton, 2007).
Although generalization in four-factor PTSD model ﬁt from adult
male military veterans to adult female civilian samples is a notable
advance, PTSD factor analytic research to date has nevertheless
been largely embedded within the adult literature. Few studies,
focusing on model ﬁt, have used child or adolescent samples despite
strong evidence that these younger age groups experience PTSD
symptoms in response to varying trauma types (Goenjian et al.,
1995; McLeer, Deblinger, Henry, & Orvaschel, 1992; Sack, Clarke,
& Seeley, 1995). Valuable exceptions to this general trend include
Saul, Grant, and Carter’s (2008) CFA study, using a sub-sample of
1581 adolescents from the National Survey of Adolescents (NSA),
and Elhai, Ford, et al.’s (2009) study of 4023 adolescents, the full
NSA sample, both of which found support for the King et al. model.
Saul et al. (2008) found that magnitudes of factor loadings of the
King et al. model signiﬁcantly differed between boys and girls. This
evidence of factorial variance across gender groups points to the
potentially inﬂuential role that gender and other demographic vari-
ables may play as moderators of the latent structure of PTSD across
diverse groups. Of particular interest, measurement invariance and
comparative ﬁt of both the King et al. and Simms et al. models of
PTSD have not been comprehensively tested across gender groups
within the same study sample in either the adult or adolescent
Given lack of clear consensus in the factor analytic literature as
to which four-factor model better represents the latent structure of
PTSD, we did not formulate speciﬁc hypotheses concerning which
of the two models would best ﬁt the data. However, we planned
606 C. Armour et al. / Journal of Anxiety Disorders 25 (2011) 604–611
a priori to use the model that provided the best model ﬁt in all
subsequent invariance-testing analyses. Speciﬁcally, based on evi-
dence of a two-fold increase in PTSD for women (Breslau et al.,
1991; Kessler et al., 1995) and reports that women experience more
chronic forms of PTSD (Breslau & Davis, 1992; Breslau et al., 1998),
we formed the general hypothesis that girls would report higher
mean levels of symptomatic distress compared to boys (as gauged
by the item-level intercepts of the 17 PTSD symptom observed indi-
cator variables). We also planned to systematically test other model
parameters (including factor loadings, covariances, variances, and
means); however, given a lack of prior evidence or well-supported
theory relating to gender-linked differences in PTSD model param-
eters in adolescents, we did not formulate any speciﬁc hypotheses
regarding the values of these parameters.
Participants were 1572 adolescent boys and girls attending nine
secondary schools who completed a screening survey during the
1997–1998 school year. All schools were located in regions in Cen-
tral Bosnia that were heavily war-exposed during the 1992–1995
Bosnian conﬂict. The purpose of the survey was to identify appro-
priate candidates for Trauma and Grief Component Therapy, a
UNICEF-sponsored school-based intervention (Layne et al., 2001,
2008; see also Cox, Moata, Clara, & Asmundson, 2008; Kutlac et al.,
2002). Demographic characteristics of the study’s effective sample
are described below.
In the Fall of 2000, in accordance with a study protocol designed
to reduce potential stigmatization of individual “high risk” stu-
dents, 16 trained school counselors used purposive sampling to
select entire classrooms in their schools with the highest prevalence
of severely war-exposed students. The counselors were psychol-
ogists and pedagogues (the duties of pedagogues most closely
resemble those of guidance counselors and learning tutors in U.S.
high schools). The counselors used all professional tools available
to them to select classrooms, including school records (which indi-
cated whether a father or mother had been killed in the war, and
whether the family was internally displaced due to the war); dis-
cussions with other teachers, parents, and school administrators;
and information gathered in their roles as teachers and counselors
at the school (cf. Layne et al., 2008, for further details concerning
classroom preparation, obtaining informed consent of caregivers
and informed consent of students, and survey administration pro-
cedures). Approximately 100 students per school completed the
risk screening survey. Students who reported signiﬁcant current
distress were then selected for further screening and considera-
tion for specialized mental health services (cf. Layne et al., 2001,
2008; for further details). Screening survey measures and proce-
dures were approved both by an ad hoc institutional review board
(IRB) formed of Bosnian mental health professionals and a local
UNICEF ofﬁcer, and by the IRB of Brigham Young University.
The screening survey included measures of pre-war, war-time,
and post-war trauma and loss exposure, post-war adversities, fre-
quency of exposure to current trauma and loss reminders, and
current psychological distress. Measures were reviewed for cultural
appropriateness and were forward- and back-translated (between
English and the local Serbian/Bosniak Muslim/Croatian native lan-
guages) by local doctoral-level psychology students and health
professionals as described elsewhere (Layne et al., 2008). Instru-
ments relevant to the present paper are described below.
2.3.1. War Trauma Screening Inventory (WTSI)
The WTSI (Layne, Stuvland, Saltzman, Djapo, & Pynoos, 1999)is
a measure of exposure to a broad range of war-related traumatic
events. The WTSI version used contained 35 “yes”/“no” self-report
items pertaining to participants’ experiences during the 1991–1995
civil war in the Former Yugoslavia that were generated based on
ﬁeld research in post-war Bosnia and on a review of war expo-
sure instruments (e.g., Macksoud & Aber, 1996). Dimensions of war
exposure assessed by the WTSI included (1) direct physical injury,
(2) witnessing violence, (3) life threat, (4) traumatic bereavement,
(5) harm to loved ones, (6) threat to loved ones, (7) displace-
ment, and (8) separations from loved ones caused by the war (see
Layne et al., 2010). All 35 items constituted Criterion A1-qualifying
events. After completing the WTSI, students were asked to record a
brief handwritten description of their “most upsetting war-related
traumatic experience” and then to indicate whether they felt “terri-
ﬁed,” “horriﬁed,” or “helpless” during that event (in three separate
“yes”/“no” items, thereby assessing PTSD Criterion A2). Respon-
dents were subsequently instructed to rate their current PTSD
symptoms relating to this most upsetting traumatic experience
(a common PTSD assessment strategy, reviewed in Elhai, Engdahl,
et al., 2009) using the UCLA PTSD Reaction Index-Revised.
2.3.2. UCLA PTSD Reaction Index-Revised
The Reaction Index (Steinberg, Brymer, Decker, & Pynoos, 2004)
is a self-report scale of PTSD symptom frequency experienced dur-
ing the previous month. The Reaction Index version used in this
study contained 17 items corresponding to the 17 DSM-IV PTSD
diagnostic criteria. Items were measured using a ﬁve-point fre-
quency scale ranging from 0 (Never) to 4 (Almost Always). The
total scale has demonstrated good internal consistency (Cronbach’s
˛ = .87), two-week test–retest reliability (r = .75), and criterion-
referenced validity in relation to a range of distress measures (r’s
ranging from .30 to .70) in post-war Bosnian adolescents sampled
in 2001 (Layne et al., 2009).
3.1. Preparatory analyses
Of 1572 students screened, 43 students endorsed no trauma
exposure and were excluded from analyses (leaving 1529); 21 were
missing more than ﬁve Reaction Index items (leaving 1508); and 28
did not identify their gender (leaving 1480 students as the effec-
tive sample size). All but four students in the effective sample were
missing fewer than three Reaction Index items. Missing items were
found to be missing completely at random (MCAR) as measured
by Little’s MCAR
(458, N = 1480) = 479.92, p > .05. We conducted
multiple imputation with an iterative Markov chain Monte Carlo
method using the Gibbs Sampler procedure (using SPSS’s Version
17 Missing Value Analysis software) to estimate missing item-level
PTSD data as generated across ﬁve imputed datasets. In addition
to the individual PTSD items, gender was also used as a predictor
variable in the imputation to account for possible between-group
missing data variation. CFA parameter estimates were averaged
across the ﬁve datasets (see Schafer & Graham, 2002) using Mplus
5.2 software (Muthén & Muthén, 2007).
CFA analyses speciﬁed both King et al.’s (1998) and Simms et al.’s
(2002) four-factor PTSD models using the Reaction Index items.
(The last loading on each of the King model factors were ﬁxed to
1.0, and the same items in the Simms factors were also ﬁxed to
1.0, for scale identiﬁcation purposes.) All items were normally dis-
tributed, with the highest univariate skewness level being 1.55 and
C. Armour et al. / Journal of Anxiety Disorders 25 (2011) 604–611 607
the highest kurtosis level being 1.86. The CFA’s thus used standard
maximum likelihood estimation. Goodness of ﬁt indices used to
evaluate model ﬁt included the comparative ﬁt index (CFI; Bentler,
1990), Tucker–Lewis Index (TLI; Tucker & Lewis, 1973), root mean
square error of approximation (RMSEA; Steiger, 1990), and stan-
dardized root mean square residual (SRMR; Jöreskog & Sörbom,
1993). Good-ﬁtting models are generally considered to fall within
the ranges of CFI and TLI ≥ .95 (.90–.94 for adequate-ﬁtting models),
RMSEA < .06 (to .08 for adequate-ﬁtting models), and SRMR < .08 (to
.10 for adequate-ﬁtting models) (Hu & Bentler, 1999). In addition,
within-group comparisons of the King et al. (1998) vs. Simms et al.
(2002) models that involved non-nested model comparisons (i.e.,
comparing a model that is not a subset of another model) used
the Bayesian Information Criterion (BIC). A 10-point BIC difference
between models denotes a 150:1 likelihood that the model with
the lower BIC value has better ﬁt (p < .05) (Raftery, 1995).
We conducted multigroup CFAs, with residual covariances ﬁxed
to zero, to test measurement invariance using parameter equality
constraints across the boys (n = 471) and girls (n = 1009). We tested
for invariance/non-invariance across groups on factor loadings,
observed variable intercepts, observed variable residual variances,
factor variances and covariances, and factor means, following
established procedures (e.g., Gregorich, 2006; Meredith, 1993;
Meredith & Teresi, 2006).
Model A allowed groups (boys and girls) to vary on all param-
eters (i.e., testing conﬁgural invariance). Subsequent models tested
progressively more conservative restrictions, constraining particu-
lar parameter estimates to be equal across groups, as tested against
the prior step’s model (except as otherwise noted). Model B con-
strained factor loadings as equal across groups (thereby testing
metric or pattern invariance). Metric invariance is established if
the corresponding factor loadings are equivalent across groups
(Gregorich, 2006). Model C additionally constrained observed vari-
able intercepts to be equal (thereby testing strong or scalar factorial
invariance). Model D additionally constrained residual variances
to be equal (thereby testing strict factorial invariance). Additional
analyses further tested structural invariance but did not constrain
residual variances, including Model E, which additionally con-
strained factor variances and covariances to be equivalent across
gender groups (tested against Model C); and Model F, which addi-
tionally constrained factor means to be equivalent across gender
groups (tested against Model E).
Tests of statistical signiﬁcance between Models A through F
were assessed with chi-square difference tests, comparing a given
CFA model assuming equal parameter estimates across groups
(e.g., factor loadings) against a model allowing those estimates to
vary across gender groups. Comparing model ﬁt statistics between
invariance testing models is not recommended given that it leads
to inaccurate results (Fan & Sivo, 2009).
The ﬁnal effective sample of 1480 participants consisted of 68.2%
(n = 1009) girls and 31.8% (n = 471) boys. Age ranged from 14 to 21
years (M = 16.18, SD = 1.07). Most participants were in the second
year of high school (n = 549, 37.4%), followed by third year (n = 425,
29.0%), ﬁrst year (n = 332, 22.6%), and fourth year (n = 160, 10.9%)
(n = 14 not reported). The mean summed Reaction Index score of
25.19 for girls (SD = 13.81; n = 1009) was signiﬁcantly higher than
that of boys (M = 16.07; SD
= 11.50; n = 471), F(1, 1478) = 154.97,
p < .001 (Cohen’s d = .66).
Analysis of BIC indices indicated that the King et al. (1998)
model provided equally good ﬁt in both the girls and boys sub-
groups; the same was true of the Simms et al. (2002) model.
Notably, BIC differences can only be used to compare models
from within a single sample (e.g., girls) and so cannot be used
to compare models across the two gender groups. For girls, the
King et al. model ﬁt well,
(113, N = 1009) = 580.18, p < .001;
CFI = .93, TLI = .92, RMSEA = .06, SRMR = .04, BIC = 49,704.059, as
Unstandardized observed item parameter estimates for the boy and girl groups.
Loadings Intercept Residual
Items Boys Girls Boys Girls Boys Girls
Intrusive thoughts .91 .87 .99 1.61 .58 .56
Nightmares .70 .75 .66 1.19 .49 .68
Reliving trauma .65 .75 .44 .90 .36 .50
Emotional cue reactivity .93 .83 1.50 2.41 .73 .75
Physiological cue reactivity 1.00 1.00 .93 1.72 .74 .83
Avoidance of thoughts 1.04 .96 1.60 2.16 1.18 1.22
Avoidance of reminders 1.00 1.00 .92 1.46 .67 .88
Trauma-related amnesia .80 .59 .52 .73 .58 .84
Loss of interest .83 .91 .78 1.08 1.33 .95
Feeling detached 1.13 1.06 .74 1.20 .67 .89
Feeling numb 1.24 1.25 .75 1.33 .67 .64
Hopelessness 1.00 1.00 .63 1.09 .82 1.07
Difﬁculty sleeping .65 .85 .49 1.02 .46 .83
Irritable/angry .78 .89 1.15 1.69 .91 .90
Difﬁculty concentrating .70 .78 .81 1.22 .62 .80
Overly alert .57 .50 2.17 2.54 1.52 1.09
Easily startled 1.00 1.00 1.01 1.86 .67 .81
did the Simms et al. model,
(113, N = 1009) = 570.42, p < .001;
CFI = .94, TLI = .92, RMSEA = .06, SRMR = .04, BIC = 49,694.26; the BIC
difference was less than 10 units. The King et al. model ﬁt ade-
quately for the boys,
(113, N = 471) = 334.12, p < .001; CFI = .92,
TLI = .90, RMSEA = .06, SRMR = .05, BIC = 22,126.41, and was nearly
adequate for the Simms et al. model,
(113, N = 471) = 343.18,
p < .001; CFI = .91, TLI = .89, RMSEA = .07, SRMR = .05, BIC = 22,135.61
(the BIC’s again differed by fewer than 10 units). Thus, neither the
King nor the Simms model was statistically superior to the other, as
tested separately in both the girls subgroup and the boys subgroup.
Given the lack of superiority in ﬁt for either model, we conducted
our subsequent analyses using the King et al. (1998) model for two
reasons. First, the King et al. model is an older and more estab-
lished model, and second, the model more closely resembles the
latent PTSD structure proposed for the DSM-V (APA, 2010).
Measurement invariance analyses formally tested differences
between girls and boys on the structural parameters of the King
et al. (1998) four-factor model (see Tables 1 and 2). Model A, which
allowed for variation in parameter estimates across groups, ﬁt
the data adequately,
(226, N = 1480) = 913.54, p < .001; CFI = .93,
TLI = .92, RMSEA = .06, SRMR = .04. Based on difference tests (see
Table 3), analyses revealed that all parameter estimates differed
signiﬁcantly (all p’s < .05) between subgroups. Speciﬁcally, the girls
subgroup had higher observed intercepts, indicating greater PTSD
item severity than the boys. The girls also had larger residual
error variances, demonstrating greater residual error that is not
Unstandardized factor parameter estimates for the A2 and non-A2 groups.
Factors Boys Girls Boys Girls
Reexperiencing (R) .70 1.09
Avoidance (A) 1.00 1.12
Numbing (N) .35 .64
Hyperarousal (H) .77 .92
R with A .70 .91
R with N .39 .67
R with H .66 .89
A with N .41 .62
A with H .66 .76
N with H .43 .71
608 C. Armour et al. / Journal of Anxiety Disorders 25 (2011) 604–611
Comparisons of boys and girls using measurement invariance procedures.
Chi-square difference test
Models tested Difference tests
A vs. B
B vs. C
C vs. D
C vs. E
E vs. F
Note: Table presents the difference between chi-square values for the model
comparisons; degrees of freedom are in parentheses. Model A = no parameters con-
strained equal across groups; Model B = factor loadings constrained to be equal;
Model C = observed variable intercepts and factor loadings constrained to be equal;
Model D = residual variances, factor loadings and observed variable intercepts con-
strained to be equal; Model E = factor variances and covariances, factor loadings
and observed variable intercepts constrained to be equal; Model F = factor means,
factor variances and covariances, factor loadings and observed variable intercepts
constrained to be equal.
p < .05.
p < .001.
accounted for by PTSD’s factors. The girls also evidenced larger fac-
tor variances, indicating greater variation in factor scores, as well as
higher co-variances indicating larger inter-correlations for PTSD’s
factors. Results were more mixed for the item-level factor loadings.
Although the great majority of factor analytic studies suggests
that the latent structure of PTSD is comprised of four factors,
no clear consensus has yet emerged regarding which four-factor
model (King et al., 1998; Simms et al., 2002) best represents the
latent structure of PTSD. The current study ﬁt both the King et al.
(1998) and Simms et al. (2002) four-factor models to data from a
sample of Bosnian war-exposed secondary school students grouped
by gender. Results indicated that both the King et al. and Simms
et al. models provided equivalently good ﬁt for girls compared
to boys. Invariance testing revealed that the girl subgroup sig-
niﬁcantly differed from the boy subgroup in all factor structure
The initial CFA produced nearly identical model ﬁt for the girls
vs. boys subgroups in both the King et al. (1998) and Simms et al.
(2002) models. Given this ﬁnding, and the rationale that the King
et al. (1998) model is better established, has received support in the
limited adolescent factor analytic literature (Elhai, Gray, Docherty,
Kashdan, & Kose, 2007; Saul et al., 2008), and is more similar to
proposed PTSD criteria for the DSM-V (APA, 2010), we tested the
factorial invariance of the King et al. model across boys vs. girls sub-
groups. Given prior evidence that women have a two-fold greater
risk for PTSD following trauma exposure (Breslau et al., 1991;
Kessler et al., 1995) and are at greater risk for chronic forms of PTSD
compared to men (Breslau & Davis, 1992; Breslau et al., 1998), we
hypothesized that adolescent girls would report higher mean levels
of distress compared to boys as gauged by the item-level indicator
intercepts. This hypothesis was conﬁrmed, in that all 17 indicator
intercepts in the girls subgroup were signiﬁcantly higher than their
respective counterparts in the boys subgroup. More generally, our
exploratory tests of factorial invariance found systematic differ-
ences between boys and girls subgroups on all model parameters.
In addition to the higher observed variable intercepts, the girls sub-
group manifested higher residual error variances, factor variances,
and factor covariances. Comparisons of the magnitudes of the factor
loadings across groups, however, yielded mixed results.
We also found higher inter-factor covariances in the girls sub-
group, pointing to a greater degree of collinearity between the
factors in girls. This ﬁnding suggests that the latent structure of
PTSD symptom factors may be more oblique (and thus potentially
more redundant across factors) in adolescent girls compared to
4.1. Potential contributors to gender-linked differences in PTSD
The ﬁnding that the girls’ observed indicator residual error vari-
ances were signiﬁcantly higher than those in the boys subgroup
suggests that the girls’ PTSD symptom factors did not capture as
much common factor variance from the individual PTSD items. This
ﬁnding suggests a greater degree of measurement error in girls’
observed responses compared to boys – a statistical artifact that
may be traceable back to an early methodological artifact: early
research into the latent structure of PTSD was largely conducted
using adult male samples (King et al., 1998; Simms et al., 2002).
This ﬁnding raises the question of whether gender discrepancies
(at least in adolescence) exist in the current diagnostic formula-
tion of PTSD, such that its factor structure maps more closely onto
its clinical manifestations as they appear in males compared to
Although this study found evidence that gender functions as
a moderator of structural parameters of latent variable models of
PTSD in adolescents, the existing child and adolescent war trauma
literature provides mixed (and still ambiguous) evidence that gen-
der performs a related function – moderating the predictive (or
causal) links between war exposure and post-war distress (Fayyad,
Karam, Karam, Tabet, & Ghosn, 2004; Powell & Durakovi
2002; Shaw, 2003). For example, in a rare follow-up study con-
ducted with 252 Croatian school-aged children, Kuterovac-Jagodic
(2003) found that demographic variables including gender and age
played a “very modest” role in predicting children’s long-term post-
traumatic stress reactions (p. 21). In contrast, in a cross-sectional
sample of 393 war-exposed Bosnian secondary school students,
ı, and Dapic
ı (2003) found that female
gender was a risk factor for both PTSD and depressive symptoms
assessed approximately one year following the Bosnian war. Of par-
ticular interest, the authors found evidence of differential relations
as a function of predicted outcome, in that war-related loss was
the most potent predictor of PTSD symptoms, whereas female gen-
der was the most potent predictor of depressive symptoms (cf.
Layne et al., 2010). Notably, much of the extant gender-related
research has been conducted using hierarchical multiple regression
and related methods, in which gender is entered as an independent
predictor or grouping variable in relation to a distress-related out-
come, rather than examined in a formal test of interaction as tests
of “true” moderation require (cf. Tolin & Foa, 2006).
Considered more generally, reasons why females tend to report
higher levels of distress compared to males following exposure to
traumatic events are not well understood. Although several expla-
nations have been put forward, most explanations pointing to a
single source or cause (e.g., exposure to more traumatic events)
have been rejected after further study (e.g., epidemiological sur-
veys found that males report higher levels of trauma exposure;
Breslau et al., 1991; Kessler et al., 1995; Tolin & Foa, 2006). Instead,
it is more likely that gender-linked differences in PTSD are linked
to a complex array of factors, including differences in exposure
to and appraisals of different types of stress, biological develop-
ment, secondary adversities set in motion by prior trauma and loss,
and the differential use of adaptive vs. maladaptive coping strate-
gies (Compas, Connor-Smith, Saltzman, Thomsen, & Wadsworth,
2001). One possible explanation, for example, is that the higher
levels of distress (at the item level) reported by the girls in the
current study may be related to higher levels of emotionality and
the use of different coping strategies compared to their male coun-
terparts (cf. Compas, Orosan, & Grant, 1993). For example, studies
suggest that women favor the use of ‘emotion’ focused coping as
C. Armour et al. / Journal of Anxiety Disorders 25 (2011) 604–611 609
opposed to men who favor the use of ‘problem’ focused coping
(Gavranidou & Rosner, 2003). Clohessy and Ehlers (1999) reported
that, compared to men, women are more likely to rely on wish-
ful thinking, memory suppression, and disengagement (avoidant
emotion coping strategies), all of which have been shown to corre-
late positively with severe PTSD symptoms. In addition, Coyne and
Racioppo (2000) reported that emotion-focused coping (especially
avoidant emotional coping), when compared to active emotional
coping, is related to poorer mental health outcomes. Tolin and
Foa (2006) also suggest that social expectancy (e.g., women are
expected to be more vulnerable to trauma and its consequences,
whereas men are expected to be more resilient and “tough”) may
also partly account for gender differences. Further study of spe-
ciﬁc dimensions of coping in relation to PTSD, depression, grief, and
other trauma and loss-related events is warranted, given that both
problem-focused and avoidant emotion-focused coping are posi-
tively associated with PTSD symptom severity (Schnider, Elhai, &
4.2. Beyond gender: other potential moderators of PTSD
Efforts to evaluate the implications of these ﬁndings for the
factor analytic literature and for future research should consider
other potential moderators of the latent structure of PTSD. For
example, Palmieri, Weathers, et al. (2007) found that the type
or appearance of the measurement instrument employed (e.g.,
administration format) is linked to the factor structure of PTSD.
Speciﬁcally, the authors found that the Simms et al. (2002) model
provided better ﬁt for data collected via the self-report PTSD Check-
list (PCL; Weathers, Litz, Huska, & Keane, 1994), whereas the King
et al. (1998) model provided better ﬁt for data collected via the
Clinician-Administered PTSD Scale (CAPS; Blake et al., 1990). How-
ever, between-model differences in ﬁt statistics were modest in
size. With respect to gender-linked differences in PTSD, Tolin and
Foa (2006) reported that between-gender differences appear to
be more pronounced when studies employ self-report question-
naires (as in the present study) rather than standardized structured
clinical interviews. Of particular interest, Tolin and Foa observed
that some interview-based studies have failed to ﬁnd signiﬁcant
gender-linked differences, whereas studies using self-report meth-
ods have generally found higher female PTSD rates. This potential
interaction between instrument format (self-report vs. interview)
and gender raises the possibility that the speciﬁc type of mea-
surement instrument employed may be linked to both systematic
differences in prevalence rates of PTSD, as well as in the latent
structure of PTSD symptoms, observed between males and females.
Essentially, when using self-report measures, girls may be over-
reporting whereas boys may be underreporting. If this is the case,
then standard screening methods may generate disproportionately
higher rates of false positives in girls, and false negatives in boys.
The clinical implications are that a disproportionally higher rate of
girls may be selected for specialized treatments.
Other format-related features of assessment instruments may
also moderate the latent structure of PTSD symptoms. Elhai,
Palmieri, Biehn, Frueh, and Magruder (2010) found systematic dif-
ferences in the latent structure of PTSD symptoms linked to the
speciﬁc self-report CAPS (Blake et al., 1990) and format used (e.g.,
rating the frequency vs. the intensity of PTSD symptoms). Further,
the limited research to date relating to the factorial invariance
of PTSD models between ethnic groups (Hoyt & Yeater, 2010;
Marshall, 2004; Norris et al., 2001) points to the need to consider
the role of culture given the ethnically diverse region (post-war
Bosnia) in which the data were collected.
Study strengths and limitations. This study is, to our knowledge,
the ﬁrst to systematically test gender as a moderator of all relevant
structural parameters of models of PTSD symptoms. We employed
a robust latent variable modeling approach and drew from a large
sample of war-exposed adolescents that is unique in the PTSD
literature. Study limitations include the assessment of PTSD via self-
report measure rather than clinical interview. However, although
clinical interviews are generally regarded as the “gold standard”
in the clinical assessment of PTSD, diagnoses based on self-report
measures vs. clinical interviews often reach an acceptable level
of agreement (Coffey, Dansky, Falsetti, Saladin, & Brady, 1998;
Harrington & Newman, 2007). Further, the latent structure of the
King et al. (1998) model of PTSD has been supported as assessed
via both clinical interviews and self-report measures. Indeed, fac-
tor analytic research has often been conducted on data collected
via self-report measures (cf. Armour & Shevlin, 2010).
In conclusion, neither the King nor the Simms model was statis-
tically superior to the other in both subgroups indicating that they
provided equivalently good ﬁt. Invariance testing revealed that the
girl subgroup signiﬁcantly differed from the boy subgroup in all fac-
tor structure parameters of the King et al. (1998) model. A greater
degree of measurement error in girls’ observed responses com-
pared to boys indicates that the factor structure maps more closely
onto its clinical manifestations as they appear in males compared
to females possibly attributable to the fact that, the construct was
originally ‘coined’ during the Vietnam War Era with young men,
and the fact that early research on the latent structure of PTSD
involved primarily male participants (King et al., 1998; Simms et al.,
2002). The higher mean levels of distress reported by females may
be attributable to higher levels of emotionality and the use of dif-
ferent coping strategies compared to their male counterparts (cf.
Compas, Orosan, & Grant, 1993). In addition, the potential inter-
action between instrument format (self-report vs. interview) and
gender raises the possibility that the speciﬁc type of measurement
instrument employed may be linked to both systematic differences
in prevalence rates of PTSD, as well as in the latent structure of
PTSD symptoms, observed between males and females. The clinical
implications being that females may be disproportionally selected
for specialized treatments.
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