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Running Head: DISSOCIATION AND ADOLESCENT SUICIDE RISK
Adolescent suicide risk and experiences of dissociation in daily life
Vera Vine*a, Sarah E. Victorb, Harmony Mohra, Amy L. Byrda, & Stephanie D. Steppa
Author Note
aDepartment of Psychiatry, University of Pittsburgh School of Medicine, Pittsburgh, PA,
USA.
bDepartment of Psychological Sciences, Texas Tech University, Lubbock, TX, USA.
*Corresponding author. Address correspondence to Vera Vine, Department of Psychiatry,
University of Pittsburgh School of Medicine, 3811 O’Hara St., Pittsburgh, PA 15213. Email:
vinevj@upmc.edu.
Declarations of interest: none
DISSOCIATION AND ADOLESCENT SUICIDE RISK
Abstract
Dissociation is associated with risk for suicide in adults, but this link is not well studied in
adolescents, in spite of their marked suicide risk. This study assessed adolescents’ dissociative
experiences in daily life and evaluated the association between dissociative experiences and
suicide risk, including the independence of this relationship from related affective and clinical
states and demographic characteristics. Clinically referred early adolescents (N=162; aged 11-13)
were assessed via multi-informant clinical interview, questionnaires, and 4-day ecological
momentary assessment protocol. Adolescents were classified as being at elevated suicide risk
using multi-informant, multi-method reports of suicide risk behavior and/or at elevated proximal
risk using the 4-day EMA only. Suicide risk was associated with daily dissociative experiences,
and this relationship was independent of daily negative and positive affect and co-occurring
borderline personality symptoms. Gender differences emerged, such that the relationship
between daily dissociative experiences and suicide risk was only significant in adolescent girls.
Overall, findings suggest dissociation may be independently relevant to adolescent suicide risk,
above and beyond effects of psychopathology and affective disturbance, and especially in girls.
Daily dissociative experiences may help understand and detect suicide risk among early
adolescents and warrant further research.
Keywords: depersonalization, derealization, affect, ecological momentary assessment, borderline
personality disorder
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“The greatest hazard of all, losing one’s self,
can occur very quietly in the world,
as if it were nothing at all.”
-- Søren Kirkegaard, The Sickness Unto Death
1. Introduction
In the last decade, suicide has risen to the 2nd leading cause of death among adolescents
(Curtin & Heron, 2019; Heron, 2016). Anonymous surveys suggest that three quarters of
adolescents’ suicidal ideation goes undisclosed, hampering prevention efforts (Pisani et al.,
2012). In order to better understand, detect, and mitigate suicide risk, it is critical to identify risk
markers reported by vulnerable adolescents more readily. The subjective phenomenon
dissociation, with longstanding implication in suicide (Frankl, 1969; Janet, 1889; Oberndorf,
1950; Walzer, 1986), could be such a useful behavioral marker. Although the relevance of
dissociation to suicide is established, conceptual and empirical precision is lacking about the
structure of dissociation, its relevance to suicide risk in adolescents, and the incremental validity
of its relation to suicide risk in a multivariate, psychopathology-informed context.
Phenomenology and Structure of Dissociation
The Diagnostic and Statistical Manual of Mental Disorders (DSM-5) defines dissociation
broadly, as “disruptions and/or discontinuities in the normal integration of consciousness,
memory, identity, emotion, perception, body representation, motor control, and behavior”
(American Psychological Association, 2013). These disruptions or discontinuities in cognitive
processes are presumed to gives rise to subjectively perceivable feeling states, which can include
derealization, a sense of unreality or detachment from the external world, and depersonalization,
a sense of unreality or detachment from one’s mind, self, or body, which can manifest as
emotional and/or physical numbness (APA, 2013). Although dissociative presentations can be
quite apparent clinically, articulating a firm conceptual boundary around dissociative experiences
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DISSOCIATION AND ADOLESCENT SUICIDE RISK
has proven challenging for several reasons. The difficulty in describing dissociated states often
leads even formal definitions to include metaphor or first-person examples (e.g., reality as
“dream-like,” Simeon et al., 2008; “feeling dead or dead inside,” Walzer, 1968; “I am no one, I
have no self,” APA, 2013). There are also prominent factor structure inconsistencies and a dearth
of naturalistic studies. Dissociation has been most well characterized in adult samples using trait
self-report measures (see two meta-analyses: Lyssenko et al., 2018; van Ijzendoorn & Schuengel,
1996). Even in these samples, the factor structure of dissociation remains heavily contested, with
findings ranging from 1 to 4, and sometimes to 7 or more, factors (Holtgraves & Stockdale,
1997; Lyssenko et al., 2018; Ruiz et al., 2008; van Ijzendoorn & Schuengel, 1996). Studies in
adolescent samples have relied thus far also on trait dissociation measures (e.g., Xavier et al.,
2018; Kisiel & Lyons, 2001; Tolmunen et al., 2007), the most common of which appears
unidimensional (Farrington et al., 2001). Only one study has assessed dissociation using
ecological momentary assessment (EMA) methodology (Greene, 2018). Because this study
assessed a narrow variant of dissociation (i.e., “peritraumatic dissociation” during chronic trauma
exposure) and was conducted in adults, it does not clearly help understand the daily dissociative
experiences that may be related to adolescent suicide risk.
Two potentially related states, not currently considered indicative of dissociation, warrant
consideration for inclusion under the dissociation umbrella: boredom and emptiness. For
instance, Stryngaris (2016) compared both boredom and emptiness to numbness, which is clearly
established under the dissociation umbrella as an aspect of depersonalization (APA, 2013). As
Stryngaris articulates, numbness, boredom, and emptiness are all clinically salient expressions of
diffuse, difficult-to-label, absences of feeling. Boredom and emptiness have been used
interchangeably (e.g., DSM-III borderline criteria; see Klonsky, 2008), and both are correlated
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with dissociative symptoms across age groups (e.g., in the context of borderline personality
disorder; Aggen et al., 2009; Chabrol et al., 2002; Conway et al, 2012). Although they differ in
some ways (e.g. Klonsky, 2008; Price, et al., preprint), dissociation and emptiness share a
particularly blurred boundary, both conceptually and phenomenologically. Theoreticians across
decades (Kernberg, 1985; Laing, 1960; Zandersen & Parnas, 2018) have conceptualized
emptiness, much like dissociation, as driven by a disruption in identity and self-perception and
resulting in visceral and existential experiences of disembodiment and unreality. Clinically, case
studies on emptiness have observed that verbatim descriptions of emptiness by clinical patients
are imprecise and often overlap with patient reports of dissociative experiences (e.g., Elsner et
al., 2018). In questionnaires, emptiness is measured using dissociation-like items, including
“inner numbness,” “I am not real,” “out of touch with myself,” (Hazell, 1982) and “absent in my
own life” (Price et al., preprint). In sum, the dissociation umbrella—already spanning
derealization, depersonalization and numbness (APA, 2013)—may also cover boredom and,
likelier, emptiness.
Relevance to Adolescent Suicide Risk
The relevance of dissociation to suicide is generally accepted. Dissociation was so
commonly observed in connection with suicide (e.g., Frankl, 1969; Janet, 1889; Oberndorf,
1950), that it was once even considered the unconscious enactment of suicide itself (Walzer,
1968). Today, dissociation is conceptualized instead as one of several responses to chronic and
acute stress, with both normative and pathological manifestations (Şar, 2014). Elevations among
adults hospitalized for imminent suicide risk have recently led dissociation to be proposed as part
of the acute suicidal state (Galynker et al., 2016). There is not yet causal evidence linking
dissociation to suicide, but initial findings suggest that repeated suicide attempts may be
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DISSOCIATION AND ADOLESCENT SUICIDE RISK
motivated by a desire to feel something (even pain) instead of numbness and emptiness (Blasco-
Fontecilla et al., 2015). Dissociative disorders have been associated with suicide attempts in a
community sample of Turkish adult women (Şar et al., 2007) and in clinical samples above and
beyond effects of co-occurring diagnoses (Foote et al., 2008; for review see Şar, 2011).
Dimensional severity of dissociative experiences is consistently elevated among adults who have
attempted suicide, according to a meta-analysis (Calati et al., 2017). Emptiness has specifically
attracted attention in the context of the suicidal process, showing elevations in both the acute
prodrome, as well as the aftermath, of adult suicide attempts (Blasco-Fontecilla et al., 2013;
Chesley, 2003; Elsner et al., 2015; Schnyder, 1999).
For adolescents, the relevance of dissociation to suicide risk is less established. Turkish
high school students who attempted suicide reported stronger dissociative symptoms than their
peers with no history of suicidal behavior (Zoroğlu et al., 2003). In another Turkish sample,
adolescents diagnosed with a dissociative disorder were significantly likelier to report having
attempted suicide than clinical and non-clinical control youth, but this effect was not significant
after controlling for gender and depression severity (Kiliç et al., 2017). In clinical youth samples,
suicide risk and suicide history were associated with only some measures of dissociation, but not
others (Kisiel & Lyons, 2001; Orbach et al., 1995). The relevance of dissociation to adolescent
suicide risk thus requires confirmation, especially during the transition to adolescence (ages 11-
13), the period when suicide risk begins to rise (Curtin & Heron, 2019) and when dissociation
may be under-reported and/or under-assessed (Steinberg, 1996).
Dissociation and Suicide in a Transdiagnostic Landscape
The processes involved in suicide risk unfold against a complex backdrop of mechanisms
of other negative outcomes, especially psychopathology. Most maladaptive psychological
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DISSOCIATION AND ADOLESCENT SUICIDE RISK
processes confer risk for multiple different negative outcomes (Insel et al., 2010), but they may
predict each outcome for outcome-specific reasons (Nolen-Hoeksema & Watkins, 2011; Vine &
Aldao, 2014). Identifying risk processes specific to suicide is especially difficult because suicide
is associated with psychopathology pervasively, yet imprecisely (Nock et al., 2019). As Nock
and colleagues (2019) explain, the high prevalence of psychopathology among suicide decedents
(close to 95%; see Cavanagh et al., 2003) makes it hard to isolate specific psychopathological
processes related directly, uniquely, and non-spuriously to suicide. At the same time, they note
that psychopathology itself is a poor predictor of suicide, because so few cases of
psychopathology (relative to all cases of psychopathology) end in suicide.
To improve the specificity of suicide risk models, Nock and colleagues (2019) have
called for strategic use of multivariate models to account for potential psychopathology
confounds. One especially likely potential confound is borderline personality disorder (BPD).
The BPD presentation prominently features both dissociation and suicide (see Conway et al.,
2012; Scalabrini et al., 2017), and BPD is one of the most common diagnoses associated with
dissociative presentations in psychiatric and other contexts (Lyssenko et al., 2018; Şar et al.,
2003; Şar et al., 2007). Additionally, BPD symptoms appear to parsimoniously represent
common psychopathology variance across disorder types in early adolescents (Vine et al., under
review), recommending a BPD covariate as an efficient tool for factoring out broad effects of
psychopathology in this period. For these reasons, to understand the specificity of the
dissociation-suicide link among youth, it is a priority to determine its independence from co-
occurring BPD symptoms.
Furthermore, the utility of dissociation as a marker of suicide risk depends also on
differentiating dissociation from related states. Negative and positive affect play robust,
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DISSOCIATION AND ADOLESCENT SUICIDE RISK
mechanistic roles in emotional disorders (Scott et al., in press), and explain important variance in
suicide risk (Rojas et al., 2015; Yamokoski et al., 2011). In dimensional studies, dissociative
experiences correlate with elevated negative affect and reduced positive-to-negative affect ratio
(Ertubo et al., 2018; Simeon et al., 2003). Although dissociation is formally defined as a
disruption in cognitive processing (APA, 2013), its assessment relies on subjective perceptions of
the dissociated feeling state. To our knowledge, no studies have attempted to differentiate
dissociation from affective states in the context of suicide risk. To determine the viability of
considering dissociation an independent marker of adolescents’ suicide risk, it is important to
isolate adolescents’ tendencies to report a dissociated feeling state from their tendencies to report
other states.
The Current Study
We examined the relationship between early adolescents’ experiences of dissociation and
suicide risk. Importantly, the entire sample could be considered at nontrivial suicide risk, given
its adolescent age range (Heron, 2016) and presence of psychopathology (Cavanagh et al., 2003).
We identified those at heightened suicide risk relative to their also-at-risk peers based on existing
histories of suicide- or self-harm-related ideation or behavior (e.g., Paul et al., 2015; for meta-
analysis see Ribiero et al., 2016). Goals of the study were to: (1) characterize the latent structure
and prevalence of dissociation experiences in this early adolescent clinical sample; and (2)
evaluate the relationship between adolescents’ dissociation and suicide risk status and probe its
independence above and beyond effects of psychopathology and affective variables. To further
pursue calls for multivariate modeling of suicide risk (Nock et al., 2019), our final aim was (3) to
explore contextual effects of demographic characteristics on dissociation in suicide risk.
1. Methods
2.1 Subjects
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DISSOCIATION AND ADOLESCENT SUICIDE RISK
Participants were 162 clinically referred adolescents aged 11-13 (Mage=12.03 years,
SD=0.92). They were recruited, with their primary caregivers, from pediatric primary care and
ambulatory psychiatric clinics in an urban, academic hospital-based setting. To ensure
impairment of a transdiagnostic nature, adolescents were oversampled1 for emotion
dysregulation based on the maximum score (parent- or adolescent-reported) from the Personality
Assessment Inventory—Adolescent Version—Affective Instability subscale (M=13.05, SD=2.90;
scores>11 indicating clinical significance; Morey, 2007). Eligible adolescents had IQ>=70,
were free of organic neurological medical conditions, current mania, and
current psychotic episodes, and were currently receiving psychiatric or behavioral
treatment for any mood or behavior problem. Half of adolescents (47%) were female; 60%
identified as racial/ethnic minorities (41% African American; 16.7% biracial; 6% American
Indian/Alaskan Native; 4% Hispanic). Most (94%) participating caregivers were female
(Mage=39.84; SD=7.25; 48% racial/ethnic minority; 88% were biological mothers). One third
(66%) of households reported not having any employed caregivers.
2.2 Procedures
Adolescents and caregivers completed questionnaires and adolescent-focused clinical
interviews as part of a larger protocol. To minimize participant burden, adolescents and
caregivers were interviewed simultaneously by two clinicians in separate rooms, with maximum
scores for each symptom retained. Ten percent of interviews were double-scored from video,
showing high inter-rater reliability (average ICC=.88). In the week following the laboratory visit,
adolescents and caregivers each completed a brief EMA component of the study. Each
1 Oversampling was conducted such that >85% of adolescents would fall in the clinical range on PAI scores (i.e., 12
or greater), while the remaining 15% were allowed to fall anywhere on the PAI range. This strategy was designed to
produce a small comparison group of nonclinical adolescents for addressing hypotheses pertinent to the original
study providing these data. In the final sample, 89% of the adolescents fell into the clinical range (observed range
12-18), while the remaining 11% displayed a range of nonclinical scores (observed range 1-11).
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DISSOCIATION AND ADOLESCENT SUICIDE RISK
participant was prompted to complete surveys 10 times over 4 days on study-provided
smartphones. Prompts were time-based to maximize compliance and avoid school hours, and
spanned consecutive days: 2 weekdays (4pm, 8pm) and 2 weekend days (12pm, 4pm, 8pm).
Compliance rates were high, with 88.8% adolescents and 90.1% caregivers completing 8 or more
prompts. All procedures were approved by the Human Research Protection Office and conducted
in an ethical manner. Informed assent and consent were obtained from each adolescent and
caregiver, respectively.
2.3 Measures.
2.3.1 Possible indicators of daily life experiences of dissociation. Five potential
indicators of dissociation were drawn from the adolescents’ EMA. Three questions had a yes/no
format: Since [last prompt]… have you felt spaced out or numb? … have you felt as though you
were in a dream? … have you had thoughts about whether or not you even existed? Two
questions initially used a 4-point Likert scale (0=not at all; 3=a lot) and asked, During the past
15 minutes, how much have you felt …empty? and …bored? Any answer other than “not at all”
was coded as an endorsement.
2.3.2 Elevated suicide risk. Four suicide risk-relevant symptoms were drawn from both
adolescent and caregiver reports on the Depression module of the Kiddie Schedule for Affective
Disorders and Schizophrenia (K-SADS-PL; Kaufman et al 1997), a semi-structured interview for
assessing the presence and severity of affective and other psychiatric disorders in 6-18-year-olds.
The symptoms used were: recurrent thoughts of death, suicidal ideation, suicidal acts, and non-
suicidal acts. Clinicians rated each symptom on a 3-point scale (0=absent; 1=subthreshold;
2=threshold), with ratings of either 1 or 2 considered endorsements for present purposes. The
Childhood Interview for DSM-IV Borderline Personality Disorder (CI-BPD; Zanarini, 2003), a
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DISSOCIATION AND ADOLESCENT SUICIDE RISK
semi-structured interview for diagnosing adolescent BPD, provided one relevant item reported
by both the adolescent and caregiver separately: recurrent suicidal behavior, gestures, or threats,
or self-mutilation behaviors. Clinicians rated this and other CI-BPD symptoms in the past 2
years (0=absent; 1=subthreshold; 2=threshold); ratings of 1 or 2 were considered endorsements.
Suicide risk items were also drawn from the Childhood Behavior Checklist (CBCL) and
Youth Self-Report (YSR) questionnaires (Achenbach, 1991). The relevant CBCL (caregiver-
reported) items were: he/she deliberately harms self or attempts suicide and talks about killing
self. The relevant YSR (adolescent-reported) items were: I deliberately try to hurt or kill myself
and I think about killing myself. Ratings of 1 or 2 were considered endorsements. Items refer to
the past 6 months (0=not true, 1=somewhat or sometimes true, and 2=very true or often true).
Lastly, the EMA assessments provided suicide risk-related items (all dichotomous,
focused on the period since the last prompt). Two adolescent-reported EMA items asked whether
they had had thoughts about killing yourself or hurting yourself, and whether they had told
someone you were going to kill yourself or hurt yourself. A caregiver-reported EMA item asked
whether the adolescent had told someone he/she was going to kill him/herself or hurt him/herself.
A binary suicide risk composite was created to reflect the history of any suicidal or self-
harm-related ideation or behavior, per either the adolescent’s or the caregiver’s report on any
measure (i.e., clinical interviews, questionnaires, EMA).2 To provide an estimate of more
proximal suicide risk, an alternative, 4-day binary suicide risk indicator reflected only the
endorsements of the above EMA items.
2.3.2 Psychopathology and affective covariates. A nonredundant BPD severity index
was created by summing the severity of CI-BPD symptoms unrelated to either dissociation or
2 Note that most relevant items did not separate thoughts from behaviors, nor did they separate suicide-related from
non-suicidal self-harm-related thoughts or behaviors, making the composite index both the most feasible and most
appropriate.
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DISSOCIATION AND ADOLESCENT SUICIDE RISK
suicide risk (i.e., omitting dissociation, emptiness, and suicide/self-harm). The final
nonredundant BPD severity index therefore reflected symptom severity related to anger, affective
instability, efforts to avoid abandonment, impulsivity in areas besides suicidal behavior, and
unstable/intense interpersonal relationships.
Estimated average daily affect was calculated using the EMA. At each prompt,
adolescents reported how much they had felt a variety of affective states over the last 15 minutes
(0=not at all; 3=a lot). Same-valanced states were averaged at each timepoint to create averages
of negative affect (NA; sad, angry, nervous, ashamed, guilty) and positive affect (PA; happy,
relaxed, excited, energetic, proud), and scores were averaged again across timepoints to reflect
each participant’s daily NA and PA.
2.4 Analytic Plan
The above indicators were computed, and preliminary descriptive and bivariate
correlational analyses conducted in SPSS v.24 (SPSS, Inc., Chicago, IL). To account for non-
normal distributions of zero-inflated categorical indicators, subsequent analyses used the
weighted least squares mean and variance adjusted estimator in MPlus (Version 8.0.0.1; Muthén
& Muthén, 1998-2011). Main analyses proceeded in three phases, following a conventional
model-building approach. (1) First, a latent between-persons dissociation factor was identified
using confirmatory factor analysis (CFA), a two-parameter logistic item response theory (IRT)
model, and χ2 difference testing. Specifically, we began by testing the appropriateness of using all
5 potential dissociation indicators (spaced/numb, dream, exist, empty, bored) to inform a latent
dissociation construct, and then we trimmed the set of indicators as needed on the basis of fit. (2)
Second, the final latent dissociation factor was regressed on the suicide risk variables. This was
initially done without covariates, and then three psychopathology and affective covariates were
12
DISSOCIATION AND ADOLESCENT SUICIDE RISK
added to the model (nonredundant BPD symptom severity, average daily NA and PA). (3) Lastly,
models were re-run incorporating demographic characteristics (gender and minority
race/ethnicity) for a more nuanced multivariate perspective on dissociation and suicide risk.
Model fit was assessed as follows: IRT models were evaluated using difficulty and
discrimination coefficients and item characteristic curves; non-IRT models were evaluated using
convention fit indicators for structural equation models, with the following considered indicators
of good fit: non-significant χ2 likelihood ratio test; Comparative Fit Index (CFI) and/or Tucker-
Lewis Index (TLI) >=.95; and/or Root Mean Square Error of Approximation (RMSEA)<.05
(McDonald & Ho, 2002).3
3. Results
3.1 Preliminary Analyses
3.1.1 Descriptive statistics. The suicide risk composite identified 99 (61.1%) adolescents
at elevated suicide risk (n=68 by adolescent report; n=89 by caregiver report). Over half (n=58;
58.6%) were identified by both dyad members; 31 were identified by caregiver report only and
10 by adolescent report only. Of the 68 adolescents who reported elevated risk, 43 (63.2%) did
so in one measurement modality only, typically (n=40; 93.0%) during the clinical interview. The
majority of those identified were female (n=9; 81.8%) and of non-minority racial/ethnic status
(n=7; 63.6%). Frequencies of specific items composing the final composite are in Table 1 (see
also Supplement 1).
The 4-day risk indicator identified a total of 11 (6.8%) adolescents (Table 1). Of these,
most (n=9; 81.8%) were identified by adolescent report only; one was identified based on both
parties’ reports; one was identified by caregiver EMA only. The majority of those identified were
3 A deidentified dataset containing variables used for the present analyses is stored at the Open Science Framework
at [website redacted during blind review].
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DISSOCIATION AND ADOLESCENT SUICIDE RISK
female (n=9; 81.8%) and of non-minority racial/ethnic status (n=7; 63.6%). Ten of the 11
adolescents at 4-day risk were also at elevated risk according to the suicide risk composite.
Most potential dissociation items were endorsed by only a minority of participants (17-
36% of adolescents), except for boredom (endorsed by 91% of adolescents). Most adolescents
who endorsed the 4 less common items (spaced/numb, dream, exist, empty) did so only once
during the 4-day EMA (see Table 2 and table note). More detailed frequencies are in Supplement
1.
3.1.2 Bivariate correlations. Dissociation indicators except for boredom correlated with
one another significantly (rs .22 to .39; Table 2). The suicide risk composite correlated
significantly with empty, and the 4-day risk indicator correlated with spaced/numb, dream, exist,
and empty with effect sizes ranging from small to medium. Psychopathology and affective
covariates were associated as expected with most key variables. Female gender was associated
with suicide risk and several dissociation indicators, while minority race and/or ethnicity was
inversely associated with suicide risk.
3.2 Main Analyses
3.2.1 Latent structure of dissociation. Given the low frequency of most dissociation
experiences within-subjects (Table 2 note), possible dissociation indicators were dichotomized to
represent for each adolescent whether the experience was reported at all during the EMA.4 A
between-persons measurement model of dissociation informed by all five possible dichotomous
indicators had good fit overall, χ2(5)=3.25, p=.662, RMSEA=.00[.00,.09], CFI=1.00, TLI=1.03.
However, consistently with preliminary analyses, the bored indicator was not informative of the
latent construct, as indicated by nonsignificant factor loading (β=.29, SE=.15, p=.055),
4 Count-based instead of dichotomous versions dissociation indicators were tried, but these produced a poorly
fitting measurement model, χ2(2)=17.79, p<.001, RMSEA=.22, CFI=.89, TLI=0.69.
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DISSOCIATION AND ADOLESCENT SUICIDE RISK
nonsignificant variance explained (R2=.09, SE=.09, p=.914), and nonsignificant IRT parameters
(difficulty and discrimination).Results demonstrated that bored was not a useful indicator of the
latent dissociation construct at any level of dissociation severity. By contrast, the IRT model
confirmed that the other 4 potential indicators were informative of latent trait dissociation over a
range of dissociation severities. Item response patterns revealed that the empty indicator
discriminated best among adolescents with lower severity trait dissociation, and the exist
indicator discriminated best among those with higher severity dissociation. Additional
background, results, and interpretation for IRT analyses can be found in Supplement 2.
Bored was therefore dropped from the measurement model, and the resulting 4-indicator
model fit the data equally well, χ2(2)=2.34, p=.310, RMSEA=.03[.00,.16], CFI=1.00, TLI=0.99,
Δχ2(3)=0.91, p=.823. All four indicators loaded significantly and had significant variance
explained. The 4-indicator measurement model (Figure 1) was thus retained to represent
dissociation in subsequent analyses.5
3.2.2 Dissociation and suicide risk. The dissociation factor was regressed on the suicide
risk composite. This produced good fit to the data, χ2(5)=9.12, p=.105, RMSEA=.07[.11,.14],
CFI=0.96, TLI=0.92, and the path from suicide risk to dissociation was significant, β=.35,
SE=.12, p=.002. Standardized factor loadings were similar to the measurement model
(spaced/numb .70, SE=.10; dream .77, SE=.09; exist .59, SE=.13; empty .78, SE=.08; all ps< .
001).
Covariates were then added to the model using the following strategy (see Figure 2). To
distinguish dissociation from other clinically-relevant states in daily life, average NA and PA
from the EMA protocol were added as additional dependent variables, in the same position as
5 Multi-level between/within-persons model solutions were explored; however, these fit the data poorly and
indicated that all meaningful variance is at the between-persons level. This is unsurprising, given the low frequency
with which adolescents reported dissociation experiences within subjects.
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DISSOCIATION AND ADOLESCENT SUICIDE RISK
dissociation. To distinguish suicide risk from related forms of clinical impairment, nonredundant
BPD symptom severity was added as an additional independent variable, in the same position as
suicide risk. Theoretically relevant correlations were then specified: suicide risk with
nonredundant BPD symptoms; the affective states (NA, PA, and dissociation) with each other.
Initially this model fit less than adequately, χ2(14)=32.60, p=.003, RMSEA=.09[.05,.13],
CFI=0.92, TLI=0.85. Comparing observed to model-implied correlations suggested allowing two
correlations among the dissociation indicators. These were added sequentially, each time
significantly improving fit (correlating dream with exist, Δχ2(1)=6.54, p=.001; then dream with
spaced/numb, Δχ2(1)=6.54, p=.001). The resulting model (Figure 2) fit the data acceptably,
χ2(12)=22.17, p=.036, RMSEA=.07[.02,.12], CFI=0.96, TLI=0.90. Importantly, the relationship
from suicide risk to dissociation was significant despite inclusion of psychopathology and
affective covariates, β=.24, SE=.10, p=.017.6
Regressing the dissociation factor on the 4-day suicide risk indicator in place of the
suicide risk composite also provided exceptional fit, χ2(5)=4.52, p=.478, RMSEA=.00[.00, .10],
CFI=1.00, TLI=1.01, and the path from 4-day suicide risk to dissociation was similarly
significant, β=.72, SE=.15, p<.001. After adding the nonredundant BPD and affective covariates
to the model, χ2(12)=21.90, p=.039, RMSEA=.07[.02,.12], CFI=0.95, TLI=0.89, the path from 4-
day suicide risk to dissociation again remained significant, β=.37, SE=.10, p<.001.
3.2.3 Dissociation and suicide risk in the context of demographics. The main model
(Figure 2) was rerun with gender and minority racial/ethnic status added as independent
variables (correlated with suicide risk, nonredundant BPD symptoms, and one another),
predicting dissociation, NA and PA. With the suicide risk composite as the risk indicator, this
6 See online supplemental materials (Supplement 3) for testing with an additional control variable representing
adolescents’ psychopathology using the CBCL/YSR Total Problems scale, which did not contribute information
above and beyond nonredundant BPD symptoms and did not diminish the robustness of the suicide risk-dissociation
relationship.
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DISSOCIATION AND ADOLESCENT SUICIDE RISK
model fit the data well, χ2(18)=28.11, p=.060, RMSEA=.06[.00,.10], CFI=0.97, TLI=0.91 (model
coefficients in Table 3, Panel A). It was notable that with demographic characteristics in the
model, the relationship between composite suicide risk and dissociation dropped to non-
significance. A post hoc multiple-group analysis was conducted to examine whether this pathway
was moderated by female gender, which had been associated with both suicide risk and
dissociation indicators (Table 2). The multiple-group model fit the data well,
χ2(33)=39.67, p=.197, RMSEA=.05[.00,.10], CFI=0.97, TLI=0.93. Among girls, the suicide-risk-
dissociation path was significant, while among boys, the path was nonsignificant (Table 3, Panels
B and C); however, the magnitude of this group difference did not reach significance,
Wald χ2(1)=1.60, p=.206.
Using the 4-day suicide risk indicator in place of the suicide risk composite, the model
with demographics showed adequate fit, χ2(18)=27.38, p=.072, RMSEA=.06[.00,.10], CFI=0.96,
TLI=0.91, and 4-day suicide risk remained a significant predictor of dissociation, β=.33, SE=.10,
p=.001 (full coefficients in Table 3, Panel D).
4. Discussion
This study adds much-needed precision to the investigation of dissociation experiences as
correlates of adolescent suicide risk. First, our results help characterize dissociation in
adolescents’ daily lives. Among these clinically referred youth, during the sampled 4-day EMA
period, dissociation experiences appeared much less universally than boredom, but more
frequently than acute suicide-related thoughts and behaviors (Tables 1 and S1). Whereas the
discarded item bored was endorsed by 91% of adolescents, the final dissociation indicators were
endorsed by a minority (17-36%) during the 4-day EMA, and typically only once by each
endorser. Perhaps due to this imbalance, the final latent dissociation factor was not informed by
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DISSOCIATION AND ADOLESCENT SUICIDE RISK
boredom. This irrelevance of boredom echoes Stryngaris’s (2016) conclusion that boredom,
despite some similarities, is distinguishable from emptiness and numbing because of its potential
importance in signaling motivational dysfunction. The reason for the high prevalence of boredom
in our sample was unclear; boredom prevalence could have been due to being a developmentally
normative experience among 11-13-year-olds, and/or due to true motivational dysfunction
(ADHD symptoms were broadly distributed in this sample; Vine et al., under review). Future
replications could confirm the structure of dissociation among adolescents, including the
apparent relevance of emptiness but not boredom, and configural invariance across normative vs.
clinical samples.
Second, this study adds precision to the relationship of dissociation to suicide risk. We
probed the independence of that relationship in a multivariate context, accounting for
associations between relevant psychopathology and affective covariates, and we found
dissociation experiences to be reliably and robustly related with adolescents’ suicide risk. The
survival of this relationship when accounting for negative and positive affect suggests
dissociation is a distinct quality of experience related to suicide for reasons other than its
valence. Its survival when accounting for nonredundant BPD symptoms further suggests that
dissociation is uniquely relevant to suicide risk, even outside of the clinical syndrome hallmarked
by both symptoms. Supplemental analyses incorporating an additional general psychopathology
covariate (Supplement S3) further underscore the independence of the suicide-dissociation
relationship. This independence is important, given the need to move beyond psychopathology-
based models of suicide risk to identify psychological processes with unique or incremental
value for understanding suicide-specific risk, apart from risk for other maladaptive outcomes
(Nock et al., 2019). It is noteworthy that the present results replicated using the 4-day-risk
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DISSOCIATION AND ADOLESCENT SUICIDE RISK
indicator, drawn from behaviors reported during the same 4 days during which dissociation
occurred. This suggests dissociation may have incremental validity as a marker not only of
relatively distal, but of more proximal suicide risk as well. Our findings thus build on the small
body of research (Foote et al., 2008; Kiliç et al., 2017) testing whether dissociation has suicide-
specific importance, and we contribute evidence supporting this notion. Future research is
needed to fully articulate the specificity of dissociation to suicide risk, including in the context of
other relevant psychopathologies, such as posttraumatic stress disorder, panic disorder, and
psychosis (e.g., Cox & Swinson, 2002; Ford & Gomez, 2015; Justo et al., 2018; Swart et al.,
2019).
Interestingly, in the context of psychopathology and affective covariates, the latent factor
structure of dissociation changed. In the measurement model of dissociation (Figure 1), as well
as in the initial regression on suicide risk, dissociation factor loadings were similar across
indicators (.6s and .7s). To our knowledge this is the first study to consider empirically whether
emptiness belongs under the dissociation umbrella, and it yielded nuanced results that could
stimulate future research on this topic. The findings of our measurement model support the
continued consideration of emptiness as a possible facet of dissociation. At the same time, the
multivariate final model distinguishes emptiness somewhat; this model suggests that emptiness
may have even greater incremental relevance to suicide risk than the other dissociation
indicators. After psychopathology and affective covariates were added to the model (Figure 2),
the indicator empty became most informative of the latent factor (loading .97), while the other
three indicators (spaced/numb, dream, and exist) weakened. Based on this finding, the
“emptiness” question appears most informative of the latent dissociation construct in this
multivariate, suicide-specific context. Future replications are needed to confirm the reliability of
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DISSOCIATION AND ADOLESCENT SUICIDE RISK
the subjective feeling of emptiness as an especially precise marker of suicide risk in adolescent
samples. If so, “emptiness” may be fruitful to incorporate into ambulatory assessments, which
can capitalize on the widespread availability of personal technologies in adolescent life to study,
detect, and mitigate suicide risk (e.g., Torous et al. 2018, Kleiman et al., 2017; Kleiman et al.,
2019).
Importantly, incorporating demographics into the main model suppressed the
dissociation-suicide risk effect. The positive association between female gender and suicide
suggested the key suppressor was gender, so we tested moderation by gender post hoc to
determine whether the suicide risk-dissociation effect was present mainly among girls. The
moderation did not reach significance (likely a power issue), but the within-gender patterns
differed as expected: the suicide risk-dissociation path was significant in girls but not boys.
Further underscoring the presence of the effect among girls, we found no suppression by
demographics when using the 4-day risk indicator, probably because virtually all the adolescents
identified by the 4-day indicator were girls. More research is needed to interpret the present
gender findings, which could either reflect a third variable problem, or contribute real clinical
information, perhaps that dissociation in daily life has special utility for understanding suicide
risk in girls. Previous studies have reliably failed to find gender differences in the severity of
self-reported dissociative symptoms, in both adolescent (Farrington et al., 2001) and adult
populations (van Ijzendoorn & Schuengel, 1996). By contrast, in our sample, two of the
dissociation indicators were correlated with female gender (spaced out/numb, empty). Future
studies could clarify the role of gender in adolescents’ experiences of dissociation and its
relationship to suicide risk.
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DISSOCIATION AND ADOLESCENT SUICIDE RISK
Our composite variable categorizing suicide risk would not have been suited to modeling
mechanistic pathways to distinct suicide-related outcomes, but this was not our intention. Suicide
and self-injurious behavior have many important distinctions in their causes, psychological
functions, and consequences (Muehlenkamp, 2005; Whitlock et al., 2013), so it is for good
reason that mechanistic research increasingly distinguishes them. Our suicide risk composite
collapsed thoughts, ideation, plans, and acts, obscuring these distinctions that other researchers
have usefully maintained (Klonsky et al., 2018). At the same time, collapsing across these
distinctions allowed us to maximize the identification of adolescent at elevated risk, likely
reducing false negatives. Given the robustness of suicide- and self-harm-related thoughts and
actions as predictors of future suicide attempt (Ribiero et al., 2016), the composite was an
optimal strategy for sorting the already-generally-at-risk adolescents into risk levels. For the
current goal to preliminarily describe adolescents’ dissociation experiences and their incremental
relevance to suicide risk within a clinical context, this method was more than adequate. Future
studies capable of distinguishing the various self-harm-related phenomena could investigate
whether dissociation is differentially related to suicide- vs. self-harm-related experiences, and
how it relates to the critically important ideation-to-action transition (see Klonsky et al., 2018).
Results must be interpreted within the context of the limitations of an archival,
correlational, between-persons analysis. This study used data collected from a larger study
focused on adolescent emotion dysregulation, so was limited to a pre-existing set of items for
creating dissociation and suicide risk composites; future work may seek to more
comprehensively assess these constructs. Because of the infrequency of dissociation
endorsements, the present data were best served by modeling dissociation indicators as
dichotomous individual differences (e.g., each adolescent either did/did not endorse empty at
21
DISSOCIATION AND ADOLESCENT SUICIDE RISK
least once in the 4 days). Extensions of this work using EMA methods with longer/denser
sampling could incorporate within-persons modeling of dissociation, which could apply state-of-
the-art dynamic models designed for other types of affect (e.g., Scott et al., in press). Future
within-persons studies could verify whether the structure and intensity of dissociation vary
within-person over time, and how these dynamics may be clinically informative. Critically, such
studies could test three, non-mutually-exclusive characterizations of the temporal link between
dissociation and suicidal experiences: (1) that dissociation precedes suicide-related thoughts or
behavior, and may or may not have causal or mechanistic effects in elevating suicide risk or
contributing to suicidal behavior; (2) that dissociation co-occurs with suicide risk elevations,
perhaps being a facet of the acute suicidal state as recently proposed (Galynker et al., 2016); and
(3) that dissociation is a lingering after-effect, like a cognitive-affective ‘scar,’ of having
previously engaged in suicidal or related thoughts or acts.
Strengths of this study include its clinically assessed, clinically referred early adolescent
sample, with prominently elevated suicide risk and strong compliance with an EMA protocol.
This is the second study to our knowledge related to dissociation using EMA methodology
(Greene, 2018), and the first demonstration that adolescents can report on daily dissociation
experiences in real-world settings. We took care to tease apart the suicide risk-dissociation
relationship from several possible confounds, affective and clinical, while also articulating the
confounding role of adolescent gender, and we replicated our findings with an alternative, 4-day
suicide risk indicator. We have also identified needs for further replication and improvement,
especially through intensive EMA designs of longer duration and temporal density. With this
study as a first step, we encourage others to incorporate dissociation items into EMA studies of
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DISSOCIATION AND ADOLESCENT SUICIDE RISK
suicidal and at-risk youth and to consider the potential importance of dissociation in adolescents’
daily lives as a marker of elevated suicide risk.
23
DISSOCIATION AND ADOLESCENT SUICIDE RISK
Preparation of this manuscript was aided by grants from the National Institute of Mental
Health (R01 MH101088; T32 MH018951; T32 MH018269; K01 MH119216).
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DISSOCIATION AND ADOLESCENT SUICIDE RISK
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DISSOCIATION AND ADOLESCENT SUICIDE RISK
Table 1. Frequencies of variables composing the suicide risk composite indicator
Number (n) of adolescents
with elevated suicide risk,
as reportedc by:
Suicide Risk-Relevant Items: Adolescents Parents
EMAa
Suicide/self-harm thoughts 9 -
Suicide/self-harm statements 3 2
Any/either EMA item 10 2
Clinical Interview
CI-BPD suicide/self-harm behaviors 44 52
K-SADS thoughts of death 39 74
K-SADS suicidal ideation 32 41
K-SADS suicidal acts 18 17
K-SADS non-suicidal acts 27 20
K-SADS non-suicidal acts only (no death/suicide endorsed on
K-SADS) 3 1
Any/either interview 56 78
Questionnaireb
YSR/CBCL suicide/self-harm acts 16 20
YSR suicidal thoughts 20 -
CBCL suicidal statements - 23
YSR/CBCL thoughts/statements only (no acts) 6 11
Any/either questionnaire item 23 31
Note. aAt any given EMA timepoint, 134 (82.7%) to 156 (96.3) adolescents responded to the
suicide/self-harm questions; at any EMA timepoint, 113 (69.8%) to 130 (80.2%) parents
responded to the adolescent suicide/self-harm question. bNs for questionnaire items ranged from
146 to 141, depending on the item. c Adolescent and parent reports of suicide risk were correlated
positively with each other (questionnaires: r=.40, p<.001; interviews: r=.57, p<.001; EMA:
r=.20, p=.009).
32
Running Head: DISSOCIATION AND ADOLESCENT SUICIDE RISK
Table 2. Descriptive statistics and bivariate correlations among study variables
M (SD)
or N (%) 1. 2. 3. 4. 5. 6. 7. 8. 9. 10. 11.
1. Female 76 (46.9%)
2. Minority status 97 (59.9%) -.09
3. Spaced/numb 43 (26.5%) .22** -.11
4. Dream 39 (24.1%) .02 -.01 .35***
5. Exist 27 (16.7%) .01 .03 .22** .33***
6. Empty 59 (36.4%) .21** -.06 .39*** .32*** .21**
7. Bored 147 (90.7%) .00 -.13 .10 .03 .09 .11
8. Suicide risk 99 (61.1%) .24** -.19* .02 .15†.09 .26** .10
9. 4-day risk 11 (6.8%) .19* -.13 .23** .19* .27*** .26** .09 .22*
10. Daily NA 0.22 (0.30) .28*** -.27** .39*** .34*** .22** .48*** .11 .27** .30***
11. Daily PA 1.59 (0.69) -.39*** .19* -.22** -.00 -.03 -.31 -.20* -.18* -.24** -.32***
12. BPD 5.86 (2.73) .11 .16* .03 .22** .07 .16* .06 .35*** .13†.19* -.10
Note. Female is coded such that 1=female, 0=male. Minority status is coded 1=minority (i.e., African American, American
Indian/Alaskan Native, and/or biracial), 0=white. Dissociation and suicide risk indicators are coded such that 1=presence of the
construct;0=absence. BPD=nonredundant BPD symptom severity (i.e., sum of CI-BPD symptoms omitting symptoms related to
suicide and dissociation). Modal frequency of endorsements of most dissociation indicators was once: spaced/numb endorsed once by
n=17 (39.5% of endorsers); dream endorsed once by n=21 (53.8% of endorsers); exist endorsed once by n=17 (63.0% of endorsers);
empty endorsed once by n=26 (44.1% of endorsers). Bored endorsed once by n=18 (12.2% of endorsers); endorsed 4 or more times by
n=96 (65.3% of endorsers).
†p<.10. *p<.05. **p<.001. ***p<.001.
DISSOCIATION AND ADOLESCENT SUICIDE RISK
Table 3. Effects of demographic characteristics on the link between suicide risk and dissociation, in three models: (1) using the
composite suicide risk indicator in the full sample (Panel A); (2) using the composite suicide risk indicator in a multiple-group
analysis by gender (Panels B and C); and (3) using 4-day suicide risk indicator in the full sample (Panel D)
A. Full Sample (N=162) B. Girls (n=76) C. Boys (n=86) D. Full Sample w/ 4-day
Suicide Risk (N=162)
SE p
SE p
SE p
SE p
Suicide riska
Dissociation .17 .10 .091 .29 .14 .043 .04 .17 .819 .33 .10 .001
NA .11 .07 .126 .21 .12 .097 .05 .10 .583 .21 .10 .040
PA -.04 .08 .632 -.16 .10 .093 .01 .12 .939 -.15 .08 .075
BPD
Dissociation .16 .11 .148 .21 .16 .202 .11 .16 .502 .18 .10 .077
NA .17 .10 .108 .20 .19 .296 .06 .07 .432 .18 .11 .090
PA -.08 .08 .315 -.23 .10 .020 .08 .10 .437 -.07 .08 .354
Female gender
Dissociation .21 .09 .148 n/a n/a n/a n/a n/a n/a .19 .10 .049
NA .22 .07 .003 n/a n/a n/a n/a n/a n/a .20 .08 .009
PA -.36 .06 .000 n/a n/a n/a n/a n/a n/a -.34 .06 .000
w/ Suicide risk .24 .07 .000 n/a n/a n/a n/a n/a n/a .19 .09 .033
w/ BPD .11 .07 .148 n/a n/a n/a n/a n/a n/a .11 .07 .148
Minority race
Dissociation -.07 .11 .540 -.03 .14 .841 -.14 .18 .416 -.063 .10 .534
NA -.26 .08 .001 -.30 .12 .015 -.25 .12 .046 -.25 .08 .001
PA .17 .07 .023 .29 .10 .006 .08 .12 .527 .15 .07 .033
w/ Suicide risk -.19 .07 .011 -.06 .12 .621 -.27 .09 .004 -.13 .08 .096
w/ BPD .16 .08 .035 .11 .11 .329 .23 .10 .022 .15 .08 .035
Note. Female is coded such that 1=female, 0=male. Minority race is coded 1=minority (i.e., African American, American
Indian/Alaskan Native, and/or biracial), 0=white. Suicide risk indicators are coded such that 1=elevated risk; 0=not elevated risk.
BPD=nonredundant BPD symptom severity (i.e., sum of CI-BPD symptoms omitting symptoms related to suicide and dissociation).
Coefficients are standardized. Significant effects are boldfaced.
a Except as otherwise noted (Panel A), Suicide risk refers to the suicide risk composite indicator informed by EMA, clinical interview,
and questionnaires.
34
Running Head: DISSOCIATION AND ADOLESCENT SUICIDE RISK
Figure 1. Final dissociation measurement model. Model-explained variance in dissociation
indicators was significant: spaced/numb, R2 = .57(.14)***; dream, R2 = .59(.15)***; exist, R2 = .
36(.15)*; empty R2 = .52(.13)***.
*p<.05. **p<.001. ***p<.001.
1.00
Daily Life
Dissociation
“As if in a
dream”
“Spaced
out or
numb”
“Even exist?”
“Felt empty”
.42
.76(.09)***
.41
.77(.10)***
.60(.13)**
.64
.72(.09)***
.48
DISSOCIATION AND ADOLESCENT SUICIDE RISK
Figure 2. Regression of dissociation on elevated suicide risk status, accounting for related
forms of affective experience and nonredundant BPD symptoms. Model coefficients are
standardized betas with standard errors.
*p<.05. **p<.001. ***p<.001.
36
.15(.10)
.
64(.09)***
.
54(.10)***
.40(.13)**
.
97(.07)***
.24(.10)*
.
23(.06)***
-.17(.08)*
.11(.10)
-.04(.08)
-.29(.09)**
.
35(.06)***
-.30(.10)**
.
71(.08)***
.32(.15)*
.44(.15)**
Suicide Risk Daily Life
Dissociation
“As if in a
dream”
“Spaced
out or
numb”
“Even exist?”
“Felt empty”
Nonredundant
BPD Symptom
Severity
Daily NA
Daily PA
