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Numerous literature reviews have appeared over the last 15 years that have
attempted to synthesize the growing body of empirical investigations of CSA effects and
correlates (e.g., Bauserman & Rind, 1997 ; Beitchman,
Zucker, Hood, DaCosta, & Akman, 1991 ; Beitchman et al., 1992 ; Black &
DeBlassie, 1993 ; Briere
& Elliot, 1994 ; Briere & Runtz, 1993 ; Browne & Finkelhor, 1986 ;
Constantine, 1981 ; Glod, 1993 ; Jumper, 1995 ; Kendall-Tackett, Williams, &
Finkelhor, 1993 ; Kilpatrick, 1987 ; Mendel, 1995 ; Neumann, Houskamp, Pollock, &
Briere, 1996 ; Rind & Tromovitch, 1997 ; Urquiza & Capra, 1990
; Watkins & Bentovim, 1992 ). These reviews have not been unanimous in their
conclusions. Below, we examine their conclusions regarding the four commonly assumed
properties of CSA discussed previously. First we examine the qualitative literature
reviews, then the fewer and more recent quantitative (i.e., meta-analytic) reviews.
Some qualitative reviewers have been cautious regarding the issue of causality
(e.g., Bauserman & Rind, 1997 ; Beitchman et al., 1991 ; Beitchman
et al., 1992 ; Constantine, 1981 ; Kilpatrick, 1987 ), arguing that the reliable
confounding of family environment problems with CSA prevents definitive conclusions
regarding the causal role of CSA in producing maladjustment. Other reviewers, although
recognizing limitations of correlational data, have nevertheless argued that causality is
the likely state of affairs (e.g., Briere & Runtz, 1993 ; Glod, 1993 ; Urquiza &
Capra, 1990 ). Some reviewers have strongly implied that CSA causes maladjustment by
consistent use of phrases that imply causation (e.g., "effects of CSA,"
"impact of CSA") and by not addressing alternative explanations (e.g., third
variables, such as family environment) that could account for the CSA-maladjustment link
(e.g., Black & DeBlassie, 1993 ; Briere & Elliot, 1994 ; Kendall-Tackett et al.,
1993 ; Mendel, 1995 ; Watkins & Bentovim, 1992 ).
Some reviewers have concluded that CSA outcomes are variable, rather than
consistently negative (e.g., Bauserman & Rind, 1997 ; Beitchman et
al., 1991 ; Beitchman et al., 1992 ; Browne & Finkelhor, 1986 ; Constantine, 1981 ;
Kilpatrick, 1987 ).
Constantine concluded that there is no inevitable outcome or set of
reactions and that responses to CSA are mediated by nonsexual factors. Beitchman et al.
(1991) argued that the prevalence of negative outcomes may be overestimated because of
overreliance on clinical samples. Browne and Finkelhor noted that only a minority of both
sexually abused (SA) children seen by clinicians and adults with a history of CSA show
serious disturbance or psychopathology. Other reviewers, however, have implied in several
different ways that CSA effects or correlates are prevalent among persons with a history
of CSA. First, some reviewers have claimed to have written "comprehensive"
reviews of the literature or summaries of "what is currently known" (e.g.,
Briere & Elliott, 1994 ; Briere & Runtz, 1993 ; Glod, 1993 ; Urquiza & Capra,
1990 ; Watkins & Bentovim, 1992 ); their conclusion that CSA is associated with
numerous symptoms then implies that negative correlates are prevalent. Second, some
reviewers have argued that studies showing a large percentage of asymptomatic persons with
a history of CSA can be explained by factors such as insensitive measures or insufficient
time for symptoms to have developed (e.g., Briere & Elliot, 1994 ; Kendall-Tackett et
al., 1993 ). This argument implies that negative effects are prevalent, even if not yet
observed in many cases. Third, some reviewers have not discussed limitations on
generalizability from their sample of (usually clinical) studies to other CSA populations
(e.g., Black & DeBlassie, 1993 ; Kendall-Tackett et al., 1993 ; Mendel, 1995 ), again
implying that findings of negative correlates apply to the entire population of persons
with CSA experiences.
Some reviewers have concluded that the intensity of CSA outcomes varies, rather
than usually being intensely negative (e.g., Bauserman & Rind, 1997
; Beitchman et al., 1991 ; Beitchman et al., 1992 ; Browne & Finkelhor, 1986 ;
Constantine, 1981 ; Kilpatrick, 1987 ). Browne and Finkelhor noted that SA persons in
community samples tend to be either normal or only slightly impaired on psychological
measures. Constantine and Kilpatrick found that negative outcomes were often absent in SA
persons in nonclinical samples. Other reviewers, however, have implied that negative
psychological effects are frequently intense by describing the "extreme psychic
pain" ( Briere & Runtz, 1993 , p. 320) or the "pronounced deleterious
effects" ( Mendel, 1995 , p. 101) that CSA is assumed to produce. Some reviewers have
further implied the intensity of CSA effects or correlates by presenting long lists of
severe disorders (e.g., posttraumatic stress, self-mutilation) associated with CSA
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(e.g., Black & DeBlassie, 1993 ; Briere & Elliot, 1994 ; Briere &
Runtz,
1993 ; Glod, 1993 ; Kendall-Tackett et al., 1993 ; Mendel, 1995 ; Urquiza & Capra,
1990 ; Watkins & Bentovim, 1992 ).
Gender equivalence.
Several reviewers have argued that the data are insufficient to address the
issue of gender differences in outcomes (e.g., Beitchman et al., 1991 ; Beitchman et al.,
1992 ; Browne & Finkelhor, 1986 ).
Constantine (1981) concluded that girls
react more negatively than boys, attributing this difference to differences between girls'
and boys' CSA experiences. Bauserman and Rind (1997) , on the basis of
a review of college, national, and convenience samples, concluded that reactions and
outcomes for boys are more likely to be neutral or positive than for girls. Many
reviewers, however, have concluded or implied that CSA is an equivalent experience for
boys and girls in terms of its negative impact (e.g., Black & DeBlassie, 1993 ; Briere
& Runtz, 1993 ; Mendel, 1995 ; Urquiza & Capra, 1990 ; Watkins &
Bentovim,
1992 ).
Black and DeBlassie stated that CSA "has, at the very least, an equivalent
impact on males and females" (p. 128). Watkins and Bentovim claimed that one
prevalent myth about CSA is that boys are less psychologically affected than girls. Mendel
dismissed as an "exercise in futility" efforts to determine whether boys or
girls are more adversely affected by CSA, and concluded that CSA "has pronounced
deleterious effects on its victims, regardless of their gender" (p. 101).
Limitations of Qualitative Literature Reviews
The qualitative literature reviews present a mixed view of causality,
pervasiveness, intensity, and gender equivalence. This inconsistency suggests the need for
additional work in synthesizing the literature. Two other considerations also indicate
such a need: sampling biases in many of the qualitative reviews, and the vulnerability of
qualitative reviews to subjectivity and imprecision.
Qualitative literature reviews have been primarily based on clinical or legal
samples, which cannot be assumed to be representative of the population of persons with a
history of CSA (Bauserman & Rind, 1997 ; Okami, 1991 ; Rind,
1995 ). Some reviews were based exclusively or almost exclusively on clinical and legal
samples (e.g., Beitchman et al., 1991 ; Black & DeBlassie, 1993 ; Glod, 1993 ;
Kendall-Tackett et al., 1993 ; Mendel, 1995 ; Watkins & Bentovim, 1992 ). Others were
based on a majority of clinical and legal samples but included a sizable minority of
nonclinical and nonlegal samples (e.g., Beitchman et al., 1992 ; Briere & Elliott,
1994 ; Briere & Runtz, 1993 ; Browne & Finkelhor, 1986 ; Constantine, 1981 ;
Kilpatrick, 1987 ; Urquiza & Capra, 1990 ). Only one of the qualitative reviews cited
previously (Bauserman & Rind, 1997 ) included a majority of nonclinical
and nonlegal samples.
Drawing conclusions from clinical and legal samples is problematic not only
because these samples cannot be assumed to be representative of the general population,
but also because data coming from these samples are vulnerable to several biases that
threaten their validity ( Pope & Hudson, 1995 ; Rind
& Tromovitch, 1997 ). Okami (1991) studied adults who
had experienced CSA as negative, neutral, or positive. Negative responders included both
clinical and nonclinical subjects. Clinical negative responders showed substantially more
pronounced adjustment problems than nonclinical negative responders.
Okami argued that
clinical participants with negative CSA experiences constitute the negative extreme of CSA
outcomes. Pope and Hudson argued that reliance on clinical samples is problematic for
several reasons. One problem is information bias, in which clinical patients, in a search
for the causes of their problems (termed effort after meaning ), are more likely
than nonclinical participants to recall events that can be classified as CSA, thus
inflating the CSA-maladjustment relationship. Another potential bias is investigator
expectancies (cf. Rosenthal, 1977 ), in which clinical researchers who believe that CSA is
a likely cause of their patients' difficulties may transmit this expectancy to patients,
thereby increasing confirming responses. Finally, Pope and Hudson argued that causality
cannot be inferred from clinical samples because CSA and family disruption are highly
confounded in this population ( Beitchman et al., 1991 ; Ney, Fung, &
Wickett, 1994 ).
Legal samples are also likely to contain the more serious cases, limiting their
generalizability.
Qualitative reviews are entirely narrative and therefore susceptible to
reviewers' own subjective interpretations ( Jumper, 1995 ). Reviewers who are convinced
that CSA is a major cause of adult psychopathology may fall prey to confirmation bias by
noting and describing study findings indicating harmful effects but ignoring or paying
less attention to findings indicating nonnegative outcomes. For example, Mendel (1995) focused on results
from Fromuth and Burkhart's (1989) midwestern sample of males to argue that boys are
harmed by their CSA experiences but paid little attention to the southeastern sample of
males reported in the same article, for whom all CSA-adjustment correlates were
nonsignificant. In a quantitative review, the latter sample would typically have received
more weight because it had 30% more participants than the former. Even when study results
generally indicate statistically significant differences in adjustment between CSA and
control participants, summarizing this information alone is inadequate ( Rosenthal &
Rosnow, 1991 ). The sizes of these differences (i.e., effect sizes) are also important;
these effect sizes can be used to assess the intensity of CSA effects or correlates (Rind
& Tromovitch, 1997 ). Only quantitative (i.e., meta-analytic) reviews can provide this
important information.
Three recent quantitative literature reviews ( Jumper, 1995 ; Neumann et al.,
1996 ; Rind & Tromovitch, 1997 ) represent a significant advance in
assessing CSA-adjustment relations because they all (a) included a sizable proportion of
nonclinical and nonlegal samples and (b) avoided subjectivity and imprecision by using
meta-analysis. Meta-analysis is a statistical technique in which statistics from a set of
studies are converted to a common metric (e.g., standard normal deviate z s,
Cohen's d s, Pearson's r s), which are then combined into one overall
statistic that can be used to (a) infer whether one variable (e.g., CSA) is significantly
associated with another (e.g., adjustment) and (b) estimate the strength of this
association (Rind & Tromovitch, 1997 ). Common metrics such as d and
r are referred to as effect sizes and can be interpreted as assessing the size of
the difference of some attribute between two groups or the magnitude of association
between two variables. As a guideline,
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Cohen (1988)
has suggested that small, medium, and large effect sizes correspond, respectively, to d
s of .20 [small], .50 [medium], and .80 [large], and to r s of .10 [small], .30 [medium], and .50 [large]. Thus, these reviews
were well suited to examining not only whether control and SA respondents differ in
adjustment, but also to what extent they differ. Two of the reviews ( Jumper, 1995 ; Rind
& Tromovitch, 1997 ) were also able to precisely compare the genders in terms of CSA
outcomes.
Jumper (1995)
examined CSA-adjustment relations from 26 published studies with 30 samples. Of 23
samples with identified sources, 30% were clinical, 26% community, 22% student, and 22%
mixed. Thus, at least 48% of the identified samples were nonclinical and
nonlegal. Most
samples (83%) consisted of female participants. Using a weighted means approach ( Shadish
& Haddock, 1994 ), Jumper meta-analyzed effect sizes ( r s) across samples for
depression, self-esteem, and symptomatology (i.e., psychological difficulties other than
depression and self-esteem problems). The overall magnitude of the relation between CSA
and symptomatology was of medium size, r = .27. Community ( r = .29) and
clinical samples ( r = .27) were similar in magnitude, but student samples were
substantially lower ( r = .09). For self-esteem, community ( r = .34) and
clinical samples ( r = .36) were also similar, whereas student samples were much
lower ( r = -.02). For depression, the community samples ( r = .17) were
closer to student ( r = .09) than clinical samples ( r = .34).
Jumper
concluded that the student samples were anomalous, possibly because symptoms had not yet
manifested at college age. The CSA-symptomatology relation was the same for men ( r =
.29) and women ( r = ..26); the CSA-self-esteem relation, however, was lower for
men ( r = -.02) than women ( r = .24). On the basis of the symptomatology
results, which were derived from nearly twice as many samples as the self-esteem results,
Jumper concluded that SA men and women do not differ in terms of subsequent psychological
adjustment.
Neumann et al. (1996) examined CSA-adjustment relations using 38 published
studies consisting exclusively of female participants, half of which were based on
nonclinical samples. These researchers computed an overall effect size ( d ) for
each study (i.e., a study-level effect size) and then meta-analyzed them, obtaining a
small to medium weighted mean effect size ( d = .37). Using Rosenthal's (1984) formula, and
assuming a 19% CSA prevalence rate for women in the general population based on Rind
and Tromovitch's (1997) estimate, we converted this d to an r . The obtained
result ( r = .14) was considerably smaller than Jumper's estimate of r =
.27. Neumann et al. also found that the magnitude of the effect sizes differed between
nonclinical ( d = .32) and clinical ( d = .50) samples. Converting these
values to r with the procedure described previously yielded r = .12 and .19,
respectively. Thus, whereas Jumper found that community and clinical samples were similar
in terms of mean effect sizes, Neumann et al. found that nonclinical samples had a lower
mean effect size than clinical samples. This difference might be due to the fact that
Neumann et al.'s nonclinical samples included student samples (but see below). Finally,
Neumann et al. found virtually identical effect sizes for samples with a mean age of 30 or
below ( d = .39) and above 30 ( d = .40), casting doubt on Jumper's
speculation that her student results might be attributable to a lack of time for symptoms
to manifest.
Rind and Tromovitch (1997) examined CSA-outcome relations from 7
male and 7 female national probability samples from the United States, Canada, Great
Britain, and Spain. These results are especially important for estimating population
parameters because these samples were all chosen to be representative of their national
populations. Rind and Tromovitch meta-analyzed mean effect sizes from each
sample (i.e., sample-level effect sizes) separately by gender and found that the magnitude
of CSA-adjustment relations was small for both men ( r = .07) and women ( r =
.10). These mean effect sizes were not statistically different. For self-reports of CSA
effects, significantly more women (68%) reported the presence of some type of negative
effect at some point after their CSA experience than did men (42%); the size of this
difference was small to medium ( r = .23). Self-reports in Baker and Duncan's
(1985) national study in Great Britain suggested that lasting negative effects for SA
persons are rare: 13% for women and 4% for men. [*]
[* Inserted by Ipce:] In the presentation of Rind c.s. in Rotterdam, The Netherlands, December 18, 1998, they have presented the next figures from Baker & Duncan (1985):
| Baker & Dunca's (1985) questions |
Males (n-79) |
Females (n-119) |
| Permanent damage |
4% |
13% |
Harmful at the time,
but no lasting effects |
33% |
51% |
| No effect |
57% |
34% |
| Improved quality of life |
6% |
2% |
Several of the national studies also
examined third variables that might account for CSA-adjustment relations. In one study,
greater sexual activity in adulthood was confounded with CSA ( Laumann, Gagnon, Michael,
& Michaels, 1994 ). In two others ( Boney-McCoy & Finkelhor, 1995 ;
Finkelhor, Hotaling, Lewis, & Smith, 1989 ), most CSA-adjustment relations remained statistically
significant after controlling for several possible confounds. However, nonsexual abuse and
neglect variables were not held constant in these analyses, weakening any causal
interpretations because CSA often occurs along with physical abuse or emotional neglect (
Ney et al., 1994 ) and because CSA-adjustment relations have been shown to disappear when
these factors are held constant (e.g., Eckenrode, Laird, & Doris, 1993 ; Ney et al.,
1994). Finally, Rind and Tromovitch reviewed the results of another national
study that found that SA girls tended to have disruption in their family, school, and
social environments both before and after their CSA experience ( Ageton, 1988 ), weakening
causal interpretations regarding CSA effects in the general population.
Synthesis of the Quantitative Reviews
All three reviews expressed caution regarding causal inferences about
CSA-adjustment relations. Jumper (1995) noted that researchers need to differentiate
between effects related to CSA and those related to other traumatic events, and to control
for family variables. Neumann et al. (1996) argued that third
variables such as other forms of maltreatment may be responsible for the CSA-adjustment
relation, and that most studies in their review did not consider the possible role of
family dynamics. About 72% of the studies in Jumper's review were also reviewed by Neumann
et al., suggesting that most of Jumper's studies also did not consider the role of family
environment. Rind and Tromovitch (1997) found that the studies in their
review usually did not use statistical control, and when they did, it was inadequate.
Thus, a quantitative review of studies using statistical control of important potential
confounds (e.g., family environment) has yet to be done and is needed to address the issue
of causality.
Only Rind and Tromovitch's (1997) review
[Page 26]
presented data relevant to how widespread negative outcomes are in the
population of persons with a history of CSA. Their findings suggest that lasting negative
effects are rare, but these results are based on only one study ( Baker & Duncan, 1985
). These considerations point to the need for further attention to this issue.
The meta-analytic reviews were especially useful for assessing the
intensity of CSA correlates or effects, indicated by weighted mean effect sizes. Neumann
et al. (1996) and Rind and Tromovitch (1997) found that the magnitude
of the relation between CSA and adjustment in the general population is small. In
contrast, Jumper's (1995) meta-analysis of community samples suggests that the magnitude
of the CSA-adjustment relation in the general population is medium in size and equivalent
to that in the clinical population. To investigate this discrepancy, we examined the
community samples used by Jumper. For symptomatology, Jumper reported the following effect
sizes: Bagley and Ramsay (1986) , r = .13; Mullen, Romans-Clarkson, Walton, and
Herbison (1988) , r = .16; Murphy et al. (1988) , r = .13; Peters (1988) , r
= ..30; Stein, Golding, Siegel, Burnam, and Sorenson (1988) , r = .31 for the
female sample and r = .37 for the male sample. We calculated the effect sizes for
these samples and obtained, respectively, r s = .21, .16, .16, .14, .15, and .12.
Because we obtained substantially lower effect sizes in the last three samples, we asked
an expert meta-analyst to calculate these values independently; his calculations confirmed
ours. [*1]
[*1] Ralph Rosnow served as the expert meta-analyst. In an attempt to
resolve our discrepancies with Jumper, we contacted her. She informed us that her
meta-analysis came from her master's thesis and that all her data and calculations were in
storage in a different part of the country. She therefore advised us that she would be
unable to help but nevertheless suggested that we proceed with our report, mentioning that
we were unable to resolve the discrepancies with her.
We
meta-analyzed the recomputed effect sizes, obtaining a small weighted mean effect size ( r
= .15), which is consistent with the results of the other two meta-analytic reviews.
We next examined the four community samples in Jumper's meta-analysis of depression
and the three in her meta-analysis of self-esteem. Although we obtained similar effect
sizes, two of the samples used in each meta-analysis (from Hunter, 1991 ) were not valid
community samples. Hunter recruited participants through newspaper advertisements and
community notices asking for volunteers who were "sexually molested as children"
(p. 207). The recruitment method suggests a convenience sample rather than a community
sample; further, the notice wording was likely to attract volunteers who had more negative
experiences. Thus, the results of Jumper's meta-analyses of depression and self-esteem for
community samples have limited generalizability.
In sum, the quantitative reviews indicate that in the entire population of
persons with a history of CSA, the magnitude of the CSA-adjustment relation is small,
implying that CSA does not typically have intensely negative psychological effects or
correlates. The results from the Neumann et al. (1996) and Rind and
Tromovitch (1997) meta-analyses, as well as results from the recomputed meta-analysis of
Jumper's (1995) community samples, suggest that the student population is not anomalous
with respect to CSA-adjustment relations. Instead, it appears that the clinical population
is anomalous.
Gender equivalence.
Using the recomputed effect sizes for Jumper's (1995) community samples,
we recalculated the weighted mean effect sizes for male and female participants for
symptomatology and found r s = .11 and .22, respectively, compared with reported
values of r = .29 and r = .26, respectively. These revised results suggest a
sex difference. Rind and Tromovitch's (1997) meta-analysis did not
reveal a sex difference in CSA-adjustment relations (although the direction of the mean
effect sizes was consistent with greater problems for SA women), although it did show a
sex difference in self-reported effects. Each meta-analysis was based on only a small
number of male samples (Jumper used four; Rind and Tromovitch used five for
CSA-adjustment relations and three for self-reported effects). Neumann et al. (1996)
examined only female samples. The mixed results regarding CSA-adjustment relations, along
with the small number of samples used, suggest the need for a more extensive meta-analytic
examination of sex differences.
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