Attending to the role of identity explorationin self-esteem.docx

Attending to the role of identity exploration
in self-esteem: Longitudinal associations
between identity styles and two features
of self-esteem
Bart Soenens,
1
Michael D. Berzonsky,
2
and Dennis R. Papini
3
Abstract
Although research suggests an interplay between identity
development and self-esteem, most studies focused on the role
of identity
commitment and measured only level of self-esteem. This study
examined longitudinal associations between Berzonsky’s (2011)
styles
of identity exploration and two distinct features of self-esteem:
level of self-esteem and contingent self-esteem. Participants
were 167
college students (mean age ¼ 19 years; 66% female) who
completed questionnaires tapping into identity styles and
features of self-
esteem at two measurement waves separated by a 4-month
interval. Both information-oriented and normative styles were
found to
be predicted by contingent self-esteem. Follow-up analyses
demonstrated that the content of contingent self-esteem

predicting both
identity styles was different. A diffuse-avoidant identity style
was predicted mainly by low levels of self-esteem. Although we
also
observed some effects of identity styles on the self-esteem
variables, the self-esteem variables had overall a more
consistent influence
on the identity styles than the other way around.
Keywords
contingent self-esteem, identity, identity style, late adolescence,
self-esteem
Erikson (1968) recognized that how adolescents and young
adults
negotiate identity conflicts and form a sense of identity has
impor-
tant repercussions for their personality development. Given that
self-esteem has long been considered a key feature of healthy
per-
sonality development (Baumeister, Campbell, Krueger, & Vohs,
2003), considerable research has addressed associations
between
features of identity formation and self-esteem.
Marcia’s (1980) identity-status paradigm has been the basis for
much research on identity and self-esteem. Marcia (1980)
concluded

that adolescents in both the achieved (high commitment and
exp-
loration) and foreclosed (high commitment and low exploration)
sta-
tuses scored higher on self-esteem than adolescents in the
moratorium
(low commitment and high exploration) and diffusion (low
commit-
ment and exploration) statuses. More recent reviews (e.g.,
Luyckx
et al., 2013; Meeus, Iedema, Helsen, & Vollebergh, 1999)
indicate
that youth with achieved and foreclosed statuses have the
highest lev-
els of self-esteem and well-being, whereas those with a
moratorium
status reported the lowest levels. Meeus et al. (1999) concluded
that
positive well-being (including self-esteem) depends almost
exclu-
sively on the degree to which youth have formed identity
commit-
ments whereas identity exploration in the absence of
commitment

(i.e., moratorium status) is injurious to well-being and self-
esteem.
Herein we aim to address two shortcomings in research on iden-
tity and self-esteem. First, research has focused too narrowly on
the
amount or quantity of identity exploration. A more
differentiated
view on individual differences in the process of identity
exploration
might reveal important new insights. We therefore relied on
Ber-
zonsky’s (2011) social-cognitive model of identity processing
styles. Second, research has focused too narrowly on the level
or amount of positive self-esteem. In a recent, multidimensional
account of self-esteem, Heppner and Kernis (2011) argued that
it
is important to distinguish between levels of self-esteem and the
degree to which self-esteem is secure (versus insecure and
defen-
sive). The relevance of this distinction to identity research was
noted by Kroger and Marcia (2011). Few studies, however, have
examined associations between identity and both aspects of

self-
esteem simultaneously. In the present study, insecure self-
esteem
was conceptualized as contingent self-esteem. Associations
between two features of self-esteem (level of self-esteem and
con-
tingent self-esteem) and identity styles will be examined in a
short-
term longitudinal study.
Berzonsky’s identity style model
According to Berzonsky (2011), people have different identity
pro-
cessing styles and vary in how they deal with identity conflicts
and
how they process identity-relevant information. Individuals with
an
informational style engage in an open, systematic examination
of
identity options, by reflecting thoroughly about their likely
implica-
tions before forming commitments. When confronted with new
and
possibly self-discrepant information, they process it in a

relatively
flexible, unbiased fashion. Individuals with a normative style do
not
1
Ghent University, Belgium
2
Department of Psychology, State University of New York at
Cortland,
USA
3 College of Arts and Sciences, South Dakota State University,
USA
Corresponding author:
Bart Soenens, Ghent University, Henri Dunantlaan 2, Ghent
9000, Belgium.
Email: [email protected]
International Journal of
Behavioral Development
1–11
ª The Author(s) 2015
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engage in an intentional personal exploration of values and
options,
but instead internalize and rely primarily on norms and
prescrip-
tions of significant others. When confronted with new identity-
relevant information, they tend to assimilate it into already
existing
and rigidly held self-views. They easily experience discrepant
information as potentially threatening and are likely to distort,
ignore or simply dismiss self-discrepant information.
Individuals
with a diffuse-avoidant style tend to procrastinate personal
decision-making until situational demands pressure them to do
so. They hold relatively weak, unstable self-views. When
required
to react to information about potentially undesirable identity-
relevant options, they readily make temporary behavioral or
verbal

accommodations that are specific to the situations that prompted
them, which reinforces the uncertainty of their commitments.
Abundant research indicates that these identity styles relate
differentially to both the content of individuals’ identity and
their
way of coping with identity-relevant conflicts. Adolescents with
an informational style rely on active problem-solving strategies;
define themselves mainly in terms of personal attributes such as
personal values and goals; and possess intrinsic goals such as
self-development and community contribution (e.g., Berzonsky,
Cieciuch, Duriez, & Soenens, 2011; Berzonsky, Macek, &
Nurmi,
2003; Duriez, Luyckx, Soenens, & Berzonsky, 2012; Soenens,
Duriez, & Goossens, 2005). Adolescents with a normative style
score
high on need for closure, prejudice, and right-wing social-
political
views; define themselves mainly in terms of collective self-
attributes such as religion and nationality; and value goals
reflecting
conformity and conservatism (e.g., Berzonsky et al., 2003;

Duriez
et al., 2012; Soenens et al., 2005). Adolescents with a diffuse-
avoidant style define themselves mainly in terms of variable
social
attributes such as popularity and reputation and they value
hedonism
(Berzonsky et al., 2011, 2003; Soenens & Vansteenkiste, 2011).
A diffuse-avoidant style is related to a variety of maladaptive
and
counterproductive coping strategies including avoidant coping,
procrastination, and self-handicapping (e.g., Berzonsky, 2011)
A number of studies have addressed associations between the
identity styles and global level of self-esteem (Beaumont &
Zukano-
vic, 2005; Crocetti, Rubini, Berzonsky, & Meeus, 2009;
Passmore,
Fogarty, Bourke, & Baker-Evans, 2005; Phillips & Pittman,
2007;
Soenens, Berzonsky, Dunkel, Papini, & Vansteenkiste, 2011).
These
studies have shown rather consistently that self-esteem is
related posi-

tively to an informational style but negatively to a diffuse-
avoidant
style. Most studies also showed positive associations between a
nor-
mative style and global self-esteem, although some studies have
shown non-significant associations (e.g., Crocetti et al., 2009).
In stud-
ies where individuals are categorized according to their
predominant
identity style, the level of global self-esteem of individuals with
a nor-
mative style is indistinguishable from that of their informational
coun-
terparts. This is surprising because the informational and
normative
styles involve qualitatively different dynamics in terms of
information
processing, coping, and goal adoption. One possible explanation
is
that although they have similar levels of self-esteem, they differ
in
terms of the security of their self-esteem. To evaluate this
possibility,
it is important to also take into account the (in)security of self-

esteem.
Differentiating between level of self-esteem and
contingent self-esteem
It is difficult to proffer a single, consensually agreed upon
defini-
tion of self-esteem, which is a multidimensional construct that
comprises adaptive and maladaptive aspects (Heppner & Kernis,
2011). A distinction between individuals’ level of global self-
esteem and the degree to which self-esteem is fragile (versus
rel-
atively more secure) has gained prominence (Deci & Ryan,
1995;
Kernis & Paradise, 2002). Level of self-esteem, frequently
assessed with the Rosenberg (1965) scale, indicates whether
peo-
ple have a low or high sense of their overall self-worth.
Although
high scores on this scale typically indicate healthy adjustment,
it
has been argued that there may also be associated costs,
including
aggression and defensiveness (Baumeister, Smart, & Boden,

1996). Fragile self-esteem (Kernis & Paradise, 2002) has been
proposed as a means to disentangle healthy from more defensive
forms of high self-esteem.
In the current study, we measured contingent self-esteem
which, according to Heppner and Kernis (2011), represents the
core of fragile self-esteem. Because people high on contingent
self-esteem hinge their self-worth on achieving socially-
prescribed or self-imposed standards, they are sensitive to
evalua-
tion. Thus, their feelings of self-worth will vary depending on
whether or not they meet their standards. Research has shown
that
levels of global self-esteem and contingent self-esteem are dis-
tinct constructs. For instance, Kernis, Lakey, and Heppner
(2008) found moderate negative associations between the
Contin-
gent Self-Esteem Scale (CSS: Paradise & Kernis, 1999) and lev-
els of global self-esteem as measured by the Rosenberg (1965)
scale. Kernis et al. (2008) also found that individuals who
scored

high on both types of self-esteem were high in verbal defensive-
ness, suggesting they engaged in defensive maneuvers to main-
tain their self-worth.
Identity styles, level of global self-esteem, and
contingent self-esteem
We hypothesized that each of the three identity styles would
dis-
play a differentiated pattern of associations with measures of
global and contingent self-esteem. First, given that an informa-
tional style is associated with personal exploration and well-
integrated, autonomously-regulated commitments (Berzonsky,
2011; Soenens et al., 2011; Soenens & Vansteenkiste, 2011),
we hypothesized that an informational style would be associated
negatively with contingent self-esteem and positively with level
of self-esteem. Second, as discussed above individuals with
high
normative style scores have been found to have high self-
esteem.
However, given that their commitments are based on the goals
and values of significant others and that they are close-minded

and engage in maladaptive defensive maneuvers (Berzonsky,
2011; Soenens et al., 2005), we hypothesized that a normative
style would relate positively to both global and contingent
self-esteem. Third, because individuals with high diffuse-
avoidant scores have weak, unstable commitments (Berzonsky
& Ferrari, 2009), we hypothesized that a diffuse-avoidant style
would be related negatively to level of global self-esteem. It is
unclear, however, how diffuse-avoidance would relate to contin-
gent self-esteem. Because contingent self-esteem is dependent
on
enduring standards, which diffuse-avoiders lack, there may be
no
reliable association. However, because people who score high
on
diffuse-avoidance are sensitive to social cues of success, there
may
be a positive relationship. Consequently, we did not advance
spe-
cific predictions about associations between the diffuse-
avoidant
style and contingent self-esteem.

2 International Journal of Behavioral Development
The present study
We evaluated associations between the three identity styles with
measures of both global and contingent self-esteem with a two-
wave longitudinal design, which enabled us to examine the
direc-
tion of these associations. Although our reasoning thus far
implies that identity styles may affect features of self-esteem, it
is possible that these features may be predictive of variation in
the adoption of identity styles or that the relationships may be
reciprocal. For instance, a diffuse-avoidant style may not only
lead to lower levels of global self-esteem and security, it may
also
represent a consequence of the insecurity linked to low global
self-esteem. Likewise, a normative identity style may not only
lead to contingent self-worth: Adopting normative standards
may

represent a defensive response to perceived threats associated
with contingent self-esteem.
Because commitment predicts level of global self-esteem (e.g.,
Luyckx et al., 2013), we controlled for individuals’ strength of
identity commitment. We also examined the role of gender
because
there is some evidence for mean-level gender differences in
some
study variables. For instance, males have been found to score
higher
on a diffuse-avoidant style (Berzonsky, 2011) and level of self-
esteem (Kling, Hyde, Showers, & Buswell, 1999) than females.
Most research, however, indicates that structural associations
between identity styles and other variables are similar across
gender
(Berzonsky, 2011).
The conceptual model guiding this study is displayed graphi-
cally in Figure 1. The two measurement waves of this
longitudinal
study were separated by a 4-month interval. Although this is a
rel-

atively short interval, our sample consisted mainly of freshman,
university students undergoing the stressful transition to
university.
Therefore, this time period represented a meaningful interval
for
examining change and stability in identity and self-esteem.
Method
Participants and procedure
Participants were undergraduate students enrolled in
introductory
psychology at a large southern university in Tennessee, USA.
The
course meets a general education requirement and includes stu-
dents representative of freshman from all of the university’s
col-
leges and majors. At T1, 246 students participated. Three of
these 246 students were older than 30 years of age and excluded
from the study because they cannot be considered late adoles-
cents/emerging adults. Of the remaining 243 students, 167
(69%) participated at both T1 and at T2 and constituted the
focal

sample of this study. This sample was 66% female with an age
range from 18 to 26 years (M ¼ 19.21, SD ¼ 2.38). The
majority
of the participants self-identified as Caucasian (71%), 16% as
African American, 5% as Hispanic, Latino, or Mexican
American,
4% as Asian American, and 4% self-identified as other. Of the
par-
ticipants, 138 (83%) were freshmen students, 22 were sopho-
mores, 5 were juniors, and 2 were seniors.
Approval for the study was obtained from the Institutional
Review Board and participants received extra course credit.
Participants responded on a scantron sheet to a written survey
containing questions relevant to the study measures and demo-
graphic information.
A logistic regression analysis was performed to test if sample
attrition (dummy coded as drop-out ¼ 0 and retention ¼ 1) was
predicted by age, gender (dummy coded as male ¼ 1 and female
¼ 2), and all study variables at Time 1 (T1). Neither age nor
gender
in Step 1 predicted attrition, �2(2) ¼ 4.90, p ¼ .09. The study
vari-
ables on Step 2 did add to the multivariate prediction of
retention,
�2(6) ¼ 13.96, p ¼ .03. However, only level of self-esteem
uniquely predicted attrition (OR ¼ 2.06, p ¼ .003): Students
who

participated at both waves had significantly higher levels of
self-
esteem at the onset of the study (M ¼ 3.30; SD ¼ 0.60)
compared
to those who dropped out (M ¼ 3.02; SD ¼ 0.64). Nonetheless,
a
direct comparison of the correlation matrices of the study
variables
at T1 revealed no significant differences between students who
par-
ticipated twice and students who participated only at T1, �
2
(21) ¼
23.82; p ¼ 0.31.
Measures
Identity-processing styles. Participants were administered the
most
recent version of the Identity Style Inventory (ISI-5), which was
revised and validated by Berzonsky et al. (2013). Berzonsky and
colleagues (2013) established the internal structure of the scales
(via exploratory and confirmatory factor analyses) as well as
their validity. Items are rated on a 5-point scale ranging from 1
(not at all like me) to 5 (very much like me). Sample items
include: ‘I handle problems in my life by actively reflecting on

them’ for the 9-item informational scale (coefficient alpha was
.73 at Time 1; .74; at Time 2); ‘I think it is better to adopt a
firm
set of beliefs than to be open-minded,’ for the 9-item normative
scale (coefficient alpha was .71 at Time 1; .77; at Time 2); and
‘Who I am changes from situation to situation,’ for the 9-item
diffuse-avoidant scale (coefficient alpha was .83 at Time 1;
.88; at Time 2).
Global self-esteem. Level of global self-esteem was assessed
with
the Rosenberg Self-Esteem Scale (RSES: Rosenberg, 1965). The
RSES contains 10 items that participants responded to on a
4-point scale ranging from 1 (strongly disagree) to 4 (strongly
agree). A sample item is: ‘At times I think I am no good at all’
(reverse scored). Coefficient alpha was .87 at Time 1 and .86 at
Time 2.
Contingent self-esteem. Contingent self-esteem was measured
with
the 15-item Contingent Self-esteem Scale (CSS) developed by
Paradise and Kernis (1999). A sample item is: ‘An important
mea-

sure of my worth is how competently I perform.’ Items were
rated
on a 5-point scale ranging from 1 (not at all like me) to 5 (very
much
like me). Evidence for the convergent validity of scores on the
scale
is provided by Heppner and Kernis (2011; see also Kernis et al.,
2008). Alpha coefficient was .79 at both Times 1 and 2.
Strength of identity commitment. The commitment scale from
the
ISI-5 (Berzonsky et al., 2013) was used to measure the
participants’
strength of identity commitment. Participants rated the nine
items
(e.g., ‘I know basically what I believe and don’t believe’) on a
5-point scale ranging from 1 (not at all like me) to 5 (very much
like
me). Data relevant to the reliability and convergent validity of
scores on the scale is provided in Berzonsky et al. (2013).
Alpha
coefficients in the current study were .80 and .81 at Times 1 &
2,
respectively.

Soenens et al. 3
Results
Descriptive statistics and correlations
Means and standard deviations are shown in Table 1. To
examine
gender and age differences, we performed a MANOVA with
gender
as a fixed factor, with age as a covariate and with each of the
study
variables as the dependent variables. Whereas gender had a
signif-
icant multivariate effect on the study variables, Wilk’s Lambda
¼
0.78; F(12, 143) ¼ 3.35, p < .001, age did not, Wilk’s Lambda
¼
0.93; F(12, 143) ¼ 0.89, p ¼ .56. Univariate ANOVAs showed
that
gender was related to a diffuse-avoidant style at T1, F(1, 154) ¼
10.67; p < .001; �2 ¼ .07, and T2, F(1, 154) ¼ 7.65; p < .001;
�2 ¼ .05, to strength of commitment at T1, F(1, 154) ¼ 9.65;
p < .001; �

2 ¼ .06) and T2, F (1, 154) ¼ 28.62; p < .001;
�2 ¼ .16, and to self-esteem at T2, F(1, 154) ¼ 7.95; p < .001;
�2 ¼ .05. At both T1 and T2, female participants scored lower
on
a diffuse-avoidant style (M ¼ 2.01; SD ¼ 0.73 and M ¼ 2.08;
SD
¼ 0.75, respectively) than male participants (M ¼ 2.44; SD ¼
0.70 and M ¼ 2.44; SD ¼ 0.88, respectively). Also, at both
waves
females scored higher on strength of commitment (M ¼ 4.17;
SD ¼
0.65 and M ¼ 4.22; SD ¼ 0.59, respectively) than males (M ¼
3.78;
SD ¼ 0.68 and M ¼ 3.68; SD ¼ 0.60, respectively). At T2 only
and unexpectedly, female participants scored higher on self-
esteem (M ¼ 3.39; SD ¼0.53) than male participants (M ¼ 3.11;
SD ¼ 0.68). Given these findings, it was decided to statistically
control for gender in the main analyses.
The correlations in Table 1 show that each of the constructs
displayed significant stability from T1 to T2. While the rank-
order stability of most constructs was moderate to high (i.e., in
the
range between .50 and .70; Roberts & DelVecchio, 2000) and
comparable with stability coefficients obtained in previous
research (e.g., Duriez et al., 2012; Luyckx, Lens, Smits, &
Goos-
sens, 2010), the stability of an informational style was lower

com-
pared to coefficients obtained in previous longitudinal studies.
1
An information-oriented style at T1 was related positively to
com-
mitment and level of self-esteem both at T1 and T2. It was unre-
lated to contingent self-esteem within each of the waves. A
normative style at T1 was related positively to commitment and
level of self-esteem at T1 only. It was related positively to
contin-
gent self-esteem at both T1 and T2. A diffuse-avoidant style at
T1
Time 2Time 1
Information-oriented style
Normative style
Diffuse-avoidant style
Commitment
Level of self-esteem
Contingent self-esteem
Information-oriented style

Normative style
Diffuse-avoidant style
Commitment
Level of self-esteem
Contingent self-esteem
Figure 1. Conceptual model guiding the study hypotheses.
Note. Full lines represent stability coefficients. Dotted lines
represent cross-lagged effects of the identity styles on the self-
esteem variable. Dashed lines
represent cross-lagged effects of the self-esteem variables on
the identity styles.
was related negatively to commitment and level of self-esteem
at
both T1 and T2. It was related positively to contingent self-
esteem
at both T1 and T2.
Primary analyses

Structural equation modeling (SEM) with latent variables was
used
to examine the main hypotheses. Analysis of the covariance
matrices was conducted using LISREL 8.54 (Jöreskog &
Sörbom,
1996) and solutions were generated using maximum-likelihood
estimation. Each of the constructs was modeled as a latent
factor
with three indicators, that is, three randomly computed parcels
(Marsh, Hau, Balla, & Grayson, 1998). To control for gender,
each
of the indicators was regressed on gender and the
unstandardized
residuals of these regression analyses were used as indicators
for
the latent variables.
Data screening of the indicator variables indicated partial
non-normality (i.e., skewness and kurtosis) at the univariate and
multivariate level. Therefore, in all models we used the
asymptotic
covariance matrix between the indicators as input and inspected
the

Satorra-Bentler Scaled chi-square (SBS-�2, Satorra & Bentler,
1994). To evaluate model goodness of fit, the Comparative Fit
Index (CFI) and the Root Mean Square Error of Approximation
(RMSEA) were selected. According to Hu and Bentler (1999),
combined cut-off values close to .95 for CFI and .06 for
RMSEA
indicate adequate model fit. SEM-modeling proceeded in two
steps.
First, we evaluated the quality of the measurement model (i.e.,
the
relations between indicators and latent constructs) and we tested
the
assumption of longitudinal invariance through Confirmatory
Factor
Analysis (CFA). We only tested for weak invariance (i.e., invar-
iance of the factor loadings across waves) and not for stronger
forms of invariance (e.g., scalar invariance) because we did not
intend to perform a mean-level comparison of the latent
constructs
between waves. Second, having established an appropriately
fitting
measurement model, we estimated structural models testing the

hypothesized relations between the latent variables.
Measurement model. The measurement model included 12 latent
variables (information-oriented style, normative style, diffuse-
avoidant style, commitment, level of self-esteem, and
contingent
self-esteem, with each of these variables being assessed twice)
and
36 indicators. In an initial estimation of the measurement
model,
the loadings of the indicators were allowed to vary between the
two
measurement points. Also, the measurement errors of the same
indicators at different measurement points were allowed to
covary
(Burkholder & Harlow, 2003). This model showed adequate fit
to
the data, SBS-�
2
(510) ¼ 709.24, p < .001; CFI ¼ 0.97; RMSEA
¼ .05. Next, a model was estimated in which the factor loadings
were set equivalent across the two measurement points, SBS-
�
2
(528) ¼ 731.78, p < .001; CFI ¼ 0.97; RMSEA ¼ .05. The fit

of this model was not significantly worse compared to the
model
with freely varying factor loadings, �SBS-�2(18) ¼ 22.71,
p ¼ .20; �CFI ¼ .002; �RMSEA < .001, indicating that the
mea-
surement model was equivalent across measurement waves.
More-
over, all constrained factor loadings were highly significant
(p < .001), ranging from .51 to .95 (mean lambda ¼ .77). In
sum,
evidence was obtained for a reliable and longitudinally
invariant
measurement model.
Structural model. The structural model tested was a full cross-
lagged model including (a) autoregressive effects (i.e., stability
coefficients) for all constructs, (b) within-time correlations
between
all variables, and (c) cross-lagged paths between the identity
styles
and the self-esteem variables and vice versa. In order not to
exam-
ine our hypotheses in an overly conservative fashion, we
initially
tested a structural model without strength of commitment
included.

As shown in Table 1, there were moderate to strong correlations
between commitment and the other study variables. By
controlling
for the variance shared between commitment and the other study
variables already in the initial models, meaningful associations
among the study variables could remain undetected. The initial
model without commitment had an adequate fit, SBS-�2(363) ¼
491.43, p < .001; CFI ¼ 0.97; RMSEA ¼ .05. The coefficients
of
this model are provided in Table 2 (left column). As can be
seen,
an information-oriented style did not predict changes in the
self-
esteem variables. Yet, increases in an information-oriented style
were predicted by high contingent self-esteem at T1. The model
also showed reciprocal associations between a normative style
and
contingent self-esteem. A normative style at T1 predicted
increases
in contingent self-esteem and contingent self-esteem at T1
Table 1. Means, standard deviations, and correlations among
study variables.

Variable 1. 2. 3. 4. 5. 6. 7. 8. 9. 10. 11. 12. 13.
1. Gender
2. Informational style T1 .15
3. Normative style T1 .07 .17*
4. Diffuse-avoidant style T1 �.28** �.27** .07
5. Commitment T1 .30** .27** .21** �.63**
6. Level of self-esteem T1 .19* .29** .18* �.42** .44**
7. Contingent self-esteem T1 .05 .12 .27** .34** �.15* �.24**
8. Informational style T2 .09 .34** .00 �.17* .21** .14 .16*
9. Normative style T2 �.02 .12 .63** .19* .07 .04 .30** .03
10. Diffuse-avoidant style T2 �.20** �.22** .14 .53** �.40**
�.40** .16* �.28** .36**
11. Commitment T2 .42** .17* .13 �.39** .55** .50** �.07
.27** �.02 �.64**
12. Level of self-esteem T2 .26** .18* .05 �.28** .35** .61**
�.18* .13 �.08 �.44** .61**
13. Contingent self-esteem T2 �.03 .05 .22** .33** �.13
�.24** .67** .14 .26** .24** �.16* �.32**
M – 3.82 3.00 2.16 4.01 3.30 3.11 3.83 3.08 2.23 4.02 3.28 3.10
SD – 0.57 0.59 0.76 0.70 0.60 0.60 0.56 0.63 0.83 0.67 0.59
0.56
Note. Gender was coded as follows: 1 ¼ male, 2 ¼ female. The
potential range of all other variables is between 1 and 5.
Analyses are based on 167 participants. Due to
missing values on some of the variables, the ns varied between
165 and 167. *p � .05; **p < .01.
Soenens et al. 5

predicted increases in a normative style. Finally, a diffuse-
avoidant
style did not predict changes in the self-esteem variables. Yet,
increases in a diffuse-avoidant style were predicted by low self-
esteem at T1.
Next, we estimated a second model in which we controlled for
strength of commitment by allowing commitment and the
identity
styles to be correlated within each of the waves and by
including
cross-lagged paths from commitment to the self-esteem
variables
(and vice versa). This model yielded adequate fit, SBS-�2(536)
¼
757.76, p < .001; CFI ¼ 0.97; RMSEA ¼ .05. Results were gen-
erally similar to the findings obtained in the model without
com-
mitment, with the following exceptions: (a) the path from a
normative style to increases in contingent self-esteem was no
lon-

ger significant, (b) a significant path showed up between a
norma-
tive style at T1 and decreases in level of self-esteem, and (c) a
significant path showed up between a diffuse-avoidant style and
increases in contingent self-esteem. Although not central to the
present study’s hypotheses, this model also showed that level of
self-esteem predicted increases in commitment (but not the
other
way around).
Moderation by gender. To examine whether gender plays a mod-
erating role in the structural associations between the study
vari-
ables, we performed a multigroup analysis. This multigroup
analysis was performed with manifest variables rather than with
latent variables because the number of parameters to be esti-
mated in a multigroup model with latent variables exceeded the
number of participants, resulting in a non-converging solution.
Specifically, we compared a constrained model (in which asso-
ciations between the variables were equivalent across gender)
and an unconstrained model (in which associations were esti-

mated freely and were allowed to vary by gender). The uncon-
strained model did not have a better fit than the constrained
model, �X2(22) ¼ 22.71, p ¼ .42; �CFI < .001; �RMSEA <
.001, showing that gender did not moderate the structural asso-
ciations in the model.
Ancillary analyses
Whereas we anticipated that contingent self-esteem would
predict
increases in a normative style, we did not expect that it would
also
predict increases in an information-oriented style. To gain
insight in
this unexpected finding, we performed a number of exploratory
additional analyses. Inspection of the items from the contingent
self-esteem scale revealed that some items reflected the extent
to
which self-esteem depends on social standards and expectations
regarding physical attractiveness (e.g., ‘My overall feelings
about
myself are heavily influenced by what I believe other people are
saying or thinking about me’) whereas other items reflected the

extent to which self-esteem depends on personal standards (e.g.,
‘A big determinant of how much I like myself is how well I
perform
up to the standards that I have set for myself’). This distinction
seemed potentially relevant for the distinction between the
information-oriented and normative styles because it has been
argued and found that individuals high on an information-
oriented style attach importance to personal attributes to define
their identity and that individuals high on a normative style
attach
importance to expectations and standards endorsed by close
others
or even at the societal level (such as physical attractiveness)
(e.g.,
Table 2. Path coefficients in the longitudinal structural model
estimating stability and cross-lagged paths from T1 to T2.
Without control for commitment With control for commitment
Stability coefficients between T1 and T2
Information-oriented .28 [.14, .43]*** .30 [.15, .45]***
Normative .65 [.48, .83]*** .66 [.49, .83]***
Diffuse-avoidant .54 [.35, .74]*** .57 [.40, .74]***

Contingent self-esteem .71 [.56, .87]*** .71 [.56, .87]***
Level of self-esteem .61 [.44, .79]*** .56 [.40, .72]***
Commitment .45 [.27, .63]***
Cross-lagged Paths from T1 to T2
Information-oriented -> Contingent self-esteem �.05 [�.18,
.08] �.06 [�.19, .07]
Information-oriented -> Level of self-esteem .02 [�.14, .17] .10
[�.05, .25]
Normative -> Contingent self-esteem .11 [.00, .22]* .05 [�.06,
.21]
Normative -> Level of self-esteem �.09 [�.24, .06] �.17 [�.17,
.00]*
Diffuse-avoidant -> Contingent self-esteem .13 [�.02, .28]þ .28
[.02, .54]*
Diffuse-avoidant -> Level of self-esteem .01 [�.16, .17] .11
[�.18, .40]
Commitment -> Contingent self-esteem .18 [�.06, .42]
Commitment -> Level of self-esteem .19 [�.08, .46]
Contingent self-esteem -> Information-oriented .19 [.06, .32]**
.19 [.06, .32]**
Contingent self-esteem -> Normative .23 [.06, .39]** .22 [.06,
.38]**
Contingent self-esteem -> Diffuse-avoidant �.02 [�.18, .13]
�.04 [�.18, .10]
Contingent self-esteem -> Commitment .05 [�.06, .16]
Level of self-esteem -> Information-oriented .14 [�.02, .30]þ
.15 [�.01, .31]þ
Level of self-esteem -> Normative �.03 [�.19, .12] �.03 [�.18,

.12]
Level of self-esteem -> Diffuse-avoidant �.21 [�.36, �.06]**
�.20 [�.35, �.05]**
Level of self-esteem -> Commitment .32 [.17, .47]***
Note. Coefficients shown are unstandardized coefficients.
Coefficients between brackets are lower and upper values of
95% confidence intervals. Analyses are based on
165 participants. þp < .10; *p � .05; **p < .01; ***p < .001.
Berzonsky, 2011). Because a diffuse-avoidant identity style also
had a longitudinal effect on contingent self-esteem (although
only
when controlling for commitment; see Table 2), we also
explored
associations between a diffuse-avoidant style and the two types
of contingent self-esteem. It was expected that a diffuse-
avoidant
style would relate primarily to self-esteem contingent upon
social
expectations because individuals with a diffuse-avoidant style
are

said to direct their behavior and identity choices on the basis of
external and situational demands (Berzonsky, 2011).
We created separate scores for social contingent self-esteem (7
items; Cronbach’s alpha ¼ .80 at T1 and .82 at T2) and personal
contingent self-esteem (3 items; Cronbach’s alpha ¼ .56 at T1
and
.63 at T2).
2
Clearly, the modest reliability of the personal contin-
gent self-esteem scale had to do with the small number of items
included. All items had item-total correlations > .30. Then we
com-
puted correlations between these two new subscales and the
identity
styles. A differentiated pattern of associations emerged, with an
information-oriented style being related exclusively to personal
contingent self-esteem. An information-oriented style at T1 was
related positively to personal contingent self-esteem at T1 and
T2
(r ¼ .32; p < .001 and r ¼ .25; p < .001, respectively).
Similarly,
an information-oriented style at T2 was related positively to
per-

sonal contingent self-esteem at T1 and T2 (r ¼ .25; p < .001 and
r ¼ .38; p < .001, respectively). None of the correlations
between
an information-oriented style and social contingent self-esteem
reached significance (all ps > .05). A normative style was
related almost exclusively to social contingent self-esteem. A
normative style at T1 was related positively to social contingent
self-esteem at T1 and T2 (r ¼ .29; p < .001 and r ¼ .25; p <
.001, respectively). Similarly, a normative style at T2 was
related positively to social contingent self-esteem at T1 and
T2 (r ¼ .37; p < .001 and r ¼ .36; p < .001, respectively). None
of the correlations between a normative style and personal
contin-
gent self-esteem reached significance, with one exception, that
is, a
correlation between normative style T2 and personal contingent
self-esteem T1 (r ¼ .25; p < .001). Finally, a diffuse-avoidant
style
was related exclusively to social contingent self-esteem. A
diffuse-avoidant style at T1 was related positively to social
contin-
gent self-esteem at T1 and T2 (r ¼ .37; p < .001 and r ¼ .36; p
<
.001, respectively). Similarly, a diffuse-avoidant style at T2

was related positively to social contingent self-esteem at T1 and
T2 (r ¼ .25; p < .001 and r ¼ .40; p < .001, respectively). None
of the
correlations between a diffuse-avoidant style and personal
contingent
self-esteem reached significance.
Overall, these exploratory ancillary analyses suggest that, while
an information-oriented style is related uniquely to personal
contin-
gent self-esteem, a normative style and a diffuse-avoidant style
are
related primarily to social contingent self-esteem.
Discussion
Research on identity and self-esteem has focused primarily on
the
role of identity commitment at the expense of self-exploration.
This longitudinal study aimed to examine dynamics of identity
exploration, as operationalized by Berzonsky’s (2011) model of
identity processing styles, and differences in the fragility and
level of self-esteem. Each of the three identity styles was
related

to a relatively unique and in some cases unexpected pattern of
longitudinal associations with level of self-esteem and
contingent
self-esteem.
Self-esteem and an informational identity style
Consistent with previous studies (e.g., Beaumont & Zukanovic,
2005; Crocetti et al., 2009), at the cross-sectional level we
found
positive associations between an informational style and level
of
self-esteem. Contrary to prediction, at the longitudinal level
informational-style scores did not significantly predict changes
in
level of self-esteem. However, we found a marginally
significant
effect of level of self-esteem on increases in informational
scores,
which was not affected by strength of commitment. Although
this
finding should be interpreted with caution, it may suggest that
the
more people feel self-worthy, the more likely they are to
possess the

confidence and personal resources to engage deliberately in a
thor-
ough exploration of identity-relevant information. If this
finding is
reliable, it suggests that associations between an informational
style
and self-esteem may need to be interpreted in a different
direction
than is usually done in cross-sectional studies.
Unexpectedly, an informational style was predicted by changes
in contingent self-esteem. Although this finding was not
hypothe-
sized, it does comport with previous research suggesting that an
informational style may not always be autonomous regulated but
may also reflect controlled functioning. For instance, Soenens
et al. (2011) found that an informational style was associated
with
both autonomous and controlled reasons for adopting identity
com-
mitments. The current findings are consistent with this finding
because contingent self-esteem represents a controlled type of
func-

tioning (Deci & Ryan, 1995).
A more detailed analysis of the association between contingent
self-esteem and an informational style revealed that the
association
was driven uniquely by items reflecting self-esteem contingent
upon personal standards. In contrast, a normative style was
related
mainly to items reflecting self-esteem contingent upon social
stan-
dards. People high on social contingent self-esteem may be rela-
tively more alienated from personally endorsed preferences and
values (e.g., Crocker & Knight, 2005). Accordingly, this finding
suggests that the personal type of contingent self-esteem
associated
with an informational style may be relatively more benign and
per-
haps adaptive than the social type of contingent self-esteem
associ-
ated with a normative style. The association between personal
contingent self-esteem and an informational style indicates that
a

mentally effortful evaluation of identity-relevant options may
be
driven by a desire to achieve and maintain feelings of self-
worth
by accomplishing personal goals. The negative affect
experienced
when their performance fails to meet their personal standards
might
reflect an adaptive type of guilt that motivates efforts to resolve
the problem (Baumeister & Heatherton, 1996). Indeed, Lutwak,
Ferrari, and Cheek (1998) found that an informational style was
positively associated with adaptive guilt as measured by the
Test
of Self-Conscious Affect, which is positively associated with
empa-
thy, perspective taking, and informed, constructive efforts to
engage in corrective actions (Tangney, Wagner, Hill-Barlow,
Marschall, & Gramzow, 1996).
Alternatively, the failure to meet personal standards may trig-
ger needs other than enhancing self-esteem. For example, a
num-
ber of personality theories (e.g., Kelly, 1955; Epstein, 1990)

postulate that people have a need to maintain a coherent and
per-
sonally meaningful conceptual system about themselves and the
world within which they live. Actions that violate or fall short
of personal standards may create a state of cognitive dissonance
that prompts open, informed efforts to resolve the problem or
accommodate and revise self-views. Hence, the items about
Soenens et al. 7
personal standards on the Paradise and Kernis (1999) measure
may reflect concerns about self-evaluation that prompt informed
efforts to resolve self-contradictions and obtain a more accurate
understanding of oneself, rather than self-enhancement. Given
the
exploratory nature of these findings, additional research is
needed
to replicate and examine in greater detail the meaning of this

association.
Self-esteem and the normative identity style
The hypothesized associations between a normative style and
con-
tingent self-esteem were largely confirmed. Although we
initially
found evidence for reciprocal associations across time between
a
normative style and contingent self-esteem, the path from a nor-
mative style to contingent self-esteem was no longer significant
when the effect of commitment was controlled. The adoption of
normative standards in a relatively automatic fashion may be an
effort to cope with uncertainty provoked by fragile, contingent
self-esteem. Consequently, individuals with high normative
scores may easily be threatened by feedback signaling failure to
achieve these standards or by information calling into question
the
value of the standards (Kernis et al., 2008). The finding that
adoption of a normative style might be driven by contingent
self-esteem may explain why a normative style has been found

to relate to defensiveness and rigidity, confirmation-biased rea-
soning, intolerance for ambiguity, need for closure, reality-
distorting defense mechanisms, and prejudice and right-wing
authoritarianism (Berzonsky, 2011; Soenens et al., 2005). These
correlates of a normative style can perhaps be understood as
attempts to protect underlying fragile feelings of self-worth.
Research has shown that self-esteem contingent on external and
social standards is detrimental to goal pursuit, social
adjustment,
and personal well-being (e.g., Crocker & Knight, 2005; Park &
Crocker, 2008). Indeed, evidence indicates that a normative
style
is related to social maladjustment, as manifested in low
empathy
(Soenens et al., 2005) and low levels of interpersonal intimacy
(Berzonsky & Kuk, 2005). Interestingly, cross-sectional
research
shows that individuals high on a normative style typically fare
rel-
atively well in terms of personal well-being (Berzonsky, 2011).
Possibly, the vulnerability of individuals with a normative style

is
visible only when they are confronted with setbacks or threats
to
their belief system.
Also, it could be the case that the well-being costs associated
with social contingent self-esteem and a normative style only
show
up across time. For example, a normative style was found to
predict
decreases in level of self-esteem, although this finding occurred
only when controlling for commitment. Although individuals
high
on a normative style report average or even high levels of self-
worth at any given point in time (e.g., Beaumont & Zukanovic,
2005), they appear to experience decreases in levels of self-
esteem across time. It is interesting to note that this effect
showed
up only after controlling for the variance shared between a
norma-
tive style and commitment, suggesting that the self-esteem of
peo-
ple with high normative scores is buttressed by the

commitments
they have internalized. When a normative style is stripped of its
main strength (i.e., high levels of commitment), the underlying
vul-
nerability comes to the surface. Research is needed to further
exam-
ine the possibility that a normative style only has a well-being
cost
across time and under conditions where previously held commit-
ments are challenged.
Self-esteem and the diffuse-avoidant identity style
We had predicted associations between a diffuse-avoidant style
and low levels of self-esteem. The findings confirmed these
neg-
ative associations both at the cross-sectional and longitudinal
level. Surprisingly, the longitudinal analyses showed that low
lev-
els of self-esteem were predictive of diffuse-avoidant scores but
not the other way around. These findings suggest that the
adoption
of a diffuse-avoidant style may be rooted in a lack of
confidence

and a limited appreciation for one’s value as a person. The more
unworthy people feel about themselves, the more they appear to
increasingly procrastinate and avoid actively dealing with
identity-relevant conflicts and decisions. Instead, they are likely
to let their direction in life depend on situational cues (such as
hedonic pleasure; Berzonsky et al., 2011).
We did not formulate a strong hypothesis about the association
between a diffuse-avoidant style and contingent self-esteem.
Indi-
viduals with high diffuse-avoidant style scores are
hypersensitive
to external (social) demands and incentives and their self-worth
may be contingent on the attainment of those standards.
However,
their standards and commitments might be too weak and
precarious
to provide a solid foundation on which they can base their self-
esteem. Testifying to the complexity of this association, an
associ-
ation between a diffuse-avoidant style and contingent self-
esteem

only was found when the lack of (and possibly volatility of
commit-
ments) was held constant. To the extent that people have
average
and at least moderately stable levels of commitment, the
adoption
of a diffuse-avoidant style would give rise to contingent self-
worth. Follow-up analyses showed that a diffuse-avoidant style
is
related primarily to self-esteem that is contingent upon the
attain-
ment of social and external standards. This finding is consistent
with the notion that individuals with a diffuse-avoidant style are
highly sensitive to external demands and incentives (Berzonsky,
2011). Future research including a more direct assessment of
com-
mitment stability may examine whether a diffuse-avoidant style
is
only related to contingent self-esteem when identity
commitments
are at least moderately stable.
Gender differences

Because there were mean-level gender differences on some of
the
variables being investigated, we conducted a multigroup
analysis
to evaluate the possibility that gender moderated the structural
rela-
tionships between the variables. The analysis revealed the
relation-
ships were not moderated by gender. This finding is consistent
with
previous research on identity processing styles (see Berzonsky,
2011). For instance, Berzonsky and Cieciuch (2014)
investigated
relationships between identity styles, identity commitment, and
six
dimensions of psychological well-being (i.e., autonomy, life
pur-
pose, mastery, personal growth, positive relations with others,
and
self-acceptance). The pattern of structural relationships was not
moderated by the gender of the participants. Consequently, it
appears as associations between identity styles, identity
commit-

ment, and measures of well-being may be similar for male and
female participants.
Limitations
A first limitation was the relatively small sample, which may
have
limited the statistical power to detect significant effects. A
number
of theoretically expected associations were only marginally
significant, suggesting that they might become significant in
larger
samples. The limited sample size also prevented us from
examining
in greater depth differences and similarities in the cross-lagged
effects between freshmen students and non-freshmen students.
3
Second, we cannot generalize the current findings to samples
that

are more heterogeneous in terms of socio-economic background,
educational level, and ethnicity. Research on identity styles and
self-esteem needs to be conducted on larger, more
representative,
and more diverse samples before generalizable conclusions can
be made and before practical recommendations can be
formulated.
Third, although the longitudinal design of the study was a
strength,
future longitudinal research, including more than 2 waves of
assess-
ment with longer intervals between waves, may enable
researchers
to better examine reciprocal dynamics and long-term effects of
identity styles and self-esteem. This would be important also to
obtain a more dynamic picture of the interplay between identity
commitment and exploration processes which, in the current
study, were examined in a relatively static fashion because of
the
2-wave design. Fourth, the drop-out analyses showed that the
par-

ticipants scored higher on self-esteem than the dropouts at T1.
Because participants had relatively high levels of self-esteem to
begin with, there may have been less room for change in self-
esteem in our sample. As such, the selective nature of the
subsam-
ple of individuals who participated twice may have led to an
underestimation of the effects involved in level of self-esteem.
Fifth, all study variables were measured through self-report.
Although this approach is logical given that all study variables
reflect intra-individual preferences and experiences, it would be
interesting to corroborate our findings using other sources of
information such as peer or parent reports of self-esteem.
Finally,
the ancillary analyses performed on the contingent self-esteem
scale were explorative in nature and need to be replicated in
future
studies. Preferably those studies will use more elaborated mea-
sures of contingent self-esteem tapping into various domains
and
facets of the construct.

Conclusion
Testifying to the importance of identity exploration in dynamics
of self-esteem, this study showed that each of Berzonsky’s iden-
tity styles was related to a different pattern of associations with
global and contingent self-esteem, even when controlling for
com-
mitment. Overall, we found more evidence for effects of self-
esteem variables on identity styles than for effects in the
opposite
direction. This is interesting because most cross-sectional
studies
to date have modeled identity styles as predictors of self-
esteem.
The current findings suggest that, by late adolescence–early
adult-
hood, features of self-esteem have become relatively more
stable
than identity styles and are influencing identity styles rather
than
being influenced by them. Importantly, these findings do not
imply that a similar direction of effects would also be obtained
in earlier developmental periods. For instance, during early ado-

lescence, a developmental period where identity styles are
already
relevant for psychosocial adjustment (Berzonsky, Branje, &
Meeus, 2007), self-esteem might be relatively more susceptible
to change (Wigfield, Eccles, Mac Iver, Reuman, & Midgley,
1991) and might be affected more strongly by identity styles
com-
pared to late adolescence. Future longitudinal research is
needed
to confirm the present findings and to extend these findings to
other developmental periods.
Funding
The author(s) received no financial support for the research,
author-
ship, and/or publication of this article.
Notes
1. To gain more insight in the low stability of an information-
oriented style, we examined whether the stability coefficient
of this style depended on participants’ grade level at university.
If, as argued in the introduction, freshmen students experience

most changes and challenges because they are in a phase of
tran-
sition, one might expect the stability in an information-oriented
style to be lowest among freshmen students. Consistent with
this
reasoning, we found that the stability of an information-oriented
style in the subsample of freshmen (n ¼ 138, r ¼ .32) was
some-
what lower than in the subsample of the other students (sopho-
mores, juniors, and seniors, n ¼ 29, r ¼ .42). It should be
noted, however, that this difference was not statistically signif-
icant, which is probably due to the small sample size of the
non-freshmen group.
2. This distinction was supported by a Principal Components
Anal-
ysis, the results of which can be obtained from the authors upon
request.
3. Although the lack of statistical power did not allow us to
com-
pare the full cross-lagged model between freshmen and non-
freshmen, we performed a number of additional analyses to

examine the degree of equivalence between both subsamples.
First, we examined the rank-order stability of all study variables
in both subsamples. While there was a tendency for most rank-
order stability coefficients to be lower in the freshmen sample
(compared to the non-freshmen sample), the difference was sig-
nificant only for a diffuse-avoidant identity style, which was
more stable in the non-freshmen sample compared to the fresh-
men sample. Second, we addressed the degree of equivalence
between both subsamples by examining potential differences
in mean-level change in each of the study variables. A repeated
measures MANOVA with time as a within-subjects IV and with
subsample as a between-subjects IV and moderator of the effect
of time showed that neither time, Wilks’ Lambda ¼ 0.99,
F(6, 157) ¼ 0.40, p ¼ .88, nor subsample, Wilks’ Lambda ¼
0.94, F(6, 157) ¼ 1.60, p ¼ .15, nor the interaction between
time
and subsample, Wilks’ Lambda ¼ 0.96, F(6, 157) ¼ 1.02,
p ¼ .42, had an overall multivariate effect. In a final analysis,
we directly compared the correlation matrices of all study vari-
ables at both time points between the two subsamples. This
anal-
ysis showed that both correlation matrices were not statistically

different, �2(78) ¼ 82.22; p ¼ 0.35. Together, these additional
analyses suggest that there were some differences between both
subsamples, with the study variables being somewhat more open
to change during the freshman year. However, these differences
were relatively small and overall the analyses provided more
evidence for similarities than for differences between the two
subsamples.
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