Journal of Personality and Social Psychology
1986, Vol. 50, No. 4, 792-799
Copyright 1986 by chc American Psychological Association, Inc.
0022-3514/86/HJ0.75
Attributions of Personality Based on Physical
Appearance, Speech, and Handwriting
Rebecca M. Warner
David B. Sugarman
Rhode Island College
University of New Hampshire
The effect of facial appearance, speech style, and handwriting on personality attributions was examined.
The source consistency hypothesis predicted that an actor will receive consistent attributions across
all three types of information. The differential information hypothesis predicted that different personality
dimensions are used to differentiate the actors within each type of information. In a 3 X 6 multivariate
analysis of variance (MANOVA) design, each judge rated a single actor/information combination on
scales of social evaluation, intellectual evaluation, activity, potency, emotionality, and sociability. Photographs of actors were differentiated primarily in terms of positive social and intellectual evaluation;
the speech of actors was differentiated primarily along an activity dimension; and the writing of the
actors was differentiated primarily along a potency dimension. This study supported the differential
information hypothesis and suggested that these three types of information about an actor may lead
judges to use different personality dimensions.
Person perception studies have shown that observers readily
make attributions about the personality traits, abilities, and
emotions of other persons based on limited information. Three
types of information have been extensively studied: facial appearance, expressive noncontent characteristics of speech (such
as pitch, tone, and tempo), and handwriting. Numerous studies
have reported that facial features and expression influence attributions about the attractiveness, pleasantness, intellectual and
social skills, and mental health of the target person (Adams, 1977;
Berscheid & Walster, 1974; Bull & Stevens, 1979; Dion, Berscheid, & Walster, 1972; Guise, Pollans, & Turkat, 1982; Jones,
Hannson, & Phillips, 1978; Unger, Hilderbrand, & Madar, 1982).
Observers seem to apply a generally positive stereotype to persons
who are physically attractive.
There has been less consensus about the kinds of attributions
that are made based on noncontent characteristics of speech.
Pitch and tempo aifected ratings of "competence" and "benevolence" (Brown, Strong, & Rencher, 1974). Acoustic parameters
such as pitch variation, amplitude variation, and tempo correlated with ratings of pleasantness, activity, potency, and various
emotions (Scherer, 1974). Attributions are made about sex, age,
personality, social class, and ethnicity based on various speech
characteristics (Scherer & Giles, 1979). Ratings of emotionality
have been linked to speech patterns in other studies (Ostwald,
1965; Williams & Stevens, 1972). Unlike the findings on physical
attractiveness, where the results could be summarized as a positive evaluative stereotype ("what is beautiful is good"), there
does not seem to be a single principle that describes the attributions given to noncontent stylistic characteristics of speech.
There has been less interest in the study of handwriting as
expressive behavior, possibly because it seems too similar to graphology. However, there are studies that document a relation
between stylistic features of handwriting (signature size, neatness,
slant, etc.) and attributions that are made about the writer, either
by the self or by others. Judgments about self-esteem, dominance,
potency, and intellectual competence vary as a function of signature size (Aiken & Zweigenhaft, 1978; Zweigenhaft, 1977).
Some investigators report that there is a stronger relation between
the handwriting style and attributions for female writers than
for male writers (Bull & Stevens, 1979; Jorgenson, 1977).
The attributions that judges make based on appearance and
expressive behavior may not agree with judgments made by clinicians or with results from standardized psychological tests or
with self-reports of the actors about their internal states. The
issue of accuracy in person perception is a complex one (cf.
Cronbach, 1955) that is beyond the scope of this article. Here,
the issue is whether judges form consistent or consensual attributions in response to a particular expressive behavior display
such as a speech sample, and not whether their attributions are
correct. The fact that actors may engage in impression management or deception further complicates the problem of obtaining
"accurate" attributions about the actor's internal states based
on the actor's expressive behaviors (cf. Baron, 1981; Edinger &
Patterson, 1983). However, even if there is not an isomorphic
relation between physical appearance or expressive behavior display and the internal states of the actor, it may be useful to examine how particular expressive behaviors influence attributions
made by outside observers.
The analysis of multichannel communication has been addressed from two different perspectives which are relevant here.
The first perspective involves the study of expressive behavior as
an important, and often neglected, means of personality assessment (Allport, 1961). According to Hall and Lindzey (1970),
Allport and his colleagues collected extensive data on the physical
appearance and expressive behaviors of persons. In one study,
judges were asked to match the speech of an actor with other
We gratefully acknowledge the assistance of Christine Barwick, Susan
Carroll, Robert Filocco, Kathy Schneider, and Nancy Wysocki in the
data collection. Thanks to Ellen Cohn, Rodney Triplet, and an anonymous
reviewer for their comments on a draft.
Correspondence concerning this article should be addressed to Rebecca
Warner, Department of Psychology, University of New Hampshire, Durham, New Hampshire 03824.
792
ATTRIBUTIONS OF PERSONALITY
793
information about that person such as physical appearance, gait
or handwriting, personality characteristics, age, occupation, and
so forth. Allport and Cantril (1934) found that matches were
made at better than chance levels, and suggested that this was
evidence for consistency among the expressive behaviors emitted
by an individual actor. Differences in expressive behavior were
quite consistent over time for an individual (Allport & Vernon,
1933). Although there was some redundancy in the information
that was available across channels, no one channel was an exact
replica of another channel in terms of the information about
personality that it contained. In the present study, one question
that is addressed is whether judges give consistent ratings to actors
when each judge receives only one type of information about
the actor (facial appearance, voice, or handwriting).
The second question focuses on person perception, or the at-
facial appearance, does that judge use different personality or
affective dimensions to describe the actor than when the judge
is relying on noncontent vocal style, handwriting, or other channels? For instance, do judges rely primarily on an "evaluative"
dimension when judging physical appearance, and primarily on
a "potency" dimension when judging handwriting? The emphasis
here is on the consistency and differentiation of judgments rather
than on accuracy (cf. Berman et al., 1976). Rather than predicting
overall primacy of the visual channel over other channels such
tributions that observers make based on the appearance and behavior of the actor. Many studies of multichannel communication
ask how much weight is given to each channel in forming the
overall judgment, particularly when the information in different
channels is discrepant (e.g., positive verbal content paired with
negative tone of voice). Mehrabian and Ferris (1967) found that
attributions about positivity of attitude were more closely related
to facial expressions than to vocal style or verbal content. Their
formula (Overall Affect Judgment = .07 X Verbal + .38 X Vocal + .55 X Visual) provides estimates of the relative importance
of these three channels. Many other investigators have found
evidence for "visual primacy" (see DePaulo, Rosenthal, Eisenstat,
Rogers, & Finkelstein, 1978, for a review).
There is an alternate way to frame the question about multichannel communication. Several studies suggest that there may
not be a consistent set of weights that describe how observers
combine information from several channels. Instead, the relative
importance of channels may depend on the attribute being judged
(Ekman, Friesen, O'Sullivan, & Scherer, 1980). Specifically, observers might rely on different personality dimensions when they
try to "decode" visual communications than when they decode
vocal style or verbal content.
Zuckerman, Amidon, Bishop, and Pomerantz (1982) reported
that the relative importance of face and voice in judging affect
varied, depending on the type of affect that was being judged.
Tone of voice was a better source of information about dominance
and submissiveness, whereas the face provided more information
about liking or disliking. Burns and Beier (1973) also found that
information channels (vocal and visual) differed with regard to
the amount of information they conveyed about various mood
states; in particular, anxiety attributions were influenced more
by the vocal than the visual channel in their study although most
of the other moods they examined (e.g., anger, happiness, sadness)
were dominated by visual information. It is also possible that
encoders differ in the extent to which they rely on visual versus
audio channels as ways of encoding affect (Berman, Shulman,
& Marwit, 1976); and that decoders differ in their skill at decoding
particular communication channels (Rosenthal, Hall, DiMatteo,
Rogers, & Archer, 1979). Situational factors are probably important determinants of the choice of channels that are used to
communicate affect (Krauss, Apple, Morency, Wenzel, & Winton, 1981).
At this point, it may be useful to ask specific questions; for
example, when a judge is relying on visual information such as
The purpose of the present study is to see whether judges who
are provided with only one information channel (face, voice,
handwriting) can consistently differentiate the actors along several
personality dimensions (social evaluation, intellectual evaluation,
potency, activity, sociability, and emotionality). It is expected
that different dimensions will be used for each channel. Specifically, based on the studies just cited, it is predicted that there
will be consistency in the judges' attributions about both social
and intellectual evaluation given to photographs. Based on the
existing research on handwriting and personality attributions, it
is reasonable to predict that judges will give consistent ratings
as vocal or verbal, it may be reasonable to predict that the visual
channel provides relatively more information about certain dimensions of affect or personality (such as a general evaluative
dimension); but that other channels might provide more information about other personality dimensions, such as potency or
dominance.
of potency and possibly intellectual evaluation based on the
handwriting samples. The existing data on attributions based on
voice does not permit clear-cut predictions, but judgments of
emotionality and activity might be made based on noncontent
vocal style.
In the present study, three types of information (photograph,
speech sample, or handwriting) were provided about 6 actors.
Each judge saw only one actor/information combination (e.g.,
only Person 1's handwriting or Person 3's speech). The judges
rated the target on bipolar adjectives that were combined into
scales measuring various kinds of attributions about the personality or ability of the target. Three hypotheses are proposed. The
first hypothesis could be termed the source consistency or actor
consistency hypothesis. This hypothesis predicts that judges will
produce similar personality attributions for each actor regardless
of the information condition; that is, if Actor 1 "s facial appearance
is rated as high on social evaluation (warmth, cheerfulness, optimism, friendliness, attractiveness), then Actor 1's speech and
handwriting should also receive high ratings on social evaluation.
If consistent information about personality is extracted by the
judges from each of these three types of information, this might
be taken to imply that the physical appearance and behavior
display emitted by the actor contain a consistent message about
personality that is similar across all three channels (as suggested
by Allport, 1961).
A second hypothesis can be termed the differential information
hypothesis. This suggests that the pattern of judgments about
the actors would be different for each type of information. It is
proposed that judges might find it easy to form high consensus
judgments about a certain personality dimension for one type
of information and difficult to form high consensus judgments
using this dimension with other types of information; for example, judges might form high consensus judgments about ac-
794
REBECCA M. WARNER AND DAVID B. SUGARMAN
tivity only when they are reacting to speech, and not when they
are evaluating photographs and writing samples.
There could also be artifactual differences in the mean ratings
given to each type of information, for instance, a tendency to
give higher ratings of intellectual competence for writing samples
than for facial appearance. Note that this is not the same as the
differential information hypothesis.
Method
Materials
The stimuli were generated in the following manner. All the members
of one section of the introductory psychology class were brought into the
lab (N - 40). Each person was allowed time to become familiar with a
passage taken from an art history textbook, and then was tape recorded
while reading the passage out loud. This paragraph was selected in order
to standardize the speech content, and it was neutral in emotional content.
Next each person posed for four color slides (head and shoulders only).
The one slide that had the most natural looking smile and the best overall
technical quality was selected for use. This selection procedure made it
possible to standardize the facial expression; the rwnsmiling photographs
Table 1
Varimax-Rotated Factor Structures for the
13 Bipolar Adjectives
Questionnaire
development
Questionnaire
confirmation
Measure
1
II
HI
Attractive
Cheerful
Friendly
Optimistic
Warm
Knowledgahle
Competent
Responsible
Intelligent
Sensible
Strong
Sturdy
Dominant
.219
.016
.137
.164
.150
.707
.558
.738
,719
.677
.252
.079
.064
,395
.704
.650
,629
.748
.093
.271
.113
.196
.117
.039
.038
-.001
.084
.180
-.096
.122
-.187
.140
.243
.050
.106
.063
.647
m
.755
.359
JOO
.714
m
.709
.038
.160
.134
.098
.032
.091
.226
.151
II
III
.158
.062
.013
.148
.118
.646
.630
.605
.560
.607
.136
.038
.143
.148
.250
.053
.264
.023
.076
.206
.048
.092
.038
.686
m
.589
Note. Underscored numbers denote factor loadings greater than .30.
varied so much that choosing nonsmiling photographs would have resulted
The ratings questionnaire consisted of the 13 bipolar adjectives that
were chosen during the instrument development phase, 9-filler items (ad-
in much variability in expressions. It should be noted that past research
ditional bipolar adjectives that were not included in the analysis), and
suggests that smiling faces convey more information about emotionality
the Buss and Plomin (1975) EASI temperament scales (four scales, each
than nonsmiling faces; the results of this study cannot safely be generalized
containing five items, assessing emotionality, activity, sociability, and im-
to nonsmiling faces. Finally each person was asked to copy the same
pulsivity). The Buss scales were chosen because they had previously been
paragraph that was used for the speech sample onto an unlined sheet of
paper using a normal style of handwriting. The only constraint placed
used to rate other persons (rather than exclusively for self-rating), because
on this task was that the entire paragraph had to fit on one side of an
mented, and because of the content areas they covered.
the factor structure and reliability of the scales was extensively docu-
8 X 11 in. page. Thus there were three types of information for each
actor: a head and shoulders photograph, a tape recorded speech sample,
and a handwriting sample. From the 40 actors available, only the 26
1
women were retained for use as actors. Six actors were chosen by random
Judges
selection from this group of 26. Random selection was used, rather than
Judges were recruited from the introductory psychology subject pool.
A total of 404 judges participated in the study; 65% of them were women.
systematic choice of models, to represent levels of some previously determined factor such as "physical attractiveness," because the intent of
Judges were asked whether they had any previous familiarity or contact
this study was to examine an ecologically valid selection of stimuli (cf.
actors or who had missing values on the adjective rating scales were eliminated from subsequent analyses, leaving an N of 382. Initial analyses
Brunswik, 1956) rather than to rigorously control the type of information
with the actors they were rating. Any judges who were familiar with the
available to the judges. Thus, the actors do not represent extremes.2 A
6 X 3 factorial design was set up, in which 18 stimuli were included—3
types of information for 6 actors. Multivariate analysis of variance (MAN-
1
In choosing actors, the researchers deliberately avoided the strategy
OVA) and discriminant analysis were used to evaluate whether the set of
of trying to obtain actors who would be very diverse with respect to age,
ratings received by each actor, by each type of information, and by each
sex, social class or regional background, ethnicity, or educational level.
actor by information combination were significantly different.
If the models differed in sex or age, for instance, differences in the personality ratings they received could be attributable to sex or age stereo-
Questionnaire Development
The dependent variables were six summated scales based on a questionnaire consisting of bipolar adjective ratings and Likert type items.
types, and there were not enough models to systematically examine these
kinds of differences. Sex of actor was a variable that was deliberately
excluded from the design because it would have greatly complicated the
analysis, and because it was not feasible to include a large enough sample
of actors to examine both sex stereotype effects on ratings and variability
There were six dependent measures: scales measuring social evaluation,
of ratings received by individual actors within each sex. We wished to
intellectual evaluation, potency, emotionality, activity, and sociability.
The first three scales were created in an instrument development study
examine individual differences in the more usual sense. Thus, the selection
of models was restricted to Caucasian female college students.
reported in detail in the Appendix. The factor structure of the bipolar
2
The mean attractiveness ratings received by the 6 female actors ranged
adjective ratings in both the instrument development study and the main
from 1.75 to 2.31 (on a scale from 0 to 4); whereas this difference was
study are presented in Table 1. These three scales are sums of bipolar
statistically significant, F(5, 381) = 2.732, p = .019, it is clear that this
random sampling procedure resulted in selection of average attractiveness.
adjective rating items that were grouped according to factor analysis results, and the Cronbach alphas for these scales ranged from .776 to .706.
The second set of three scales were taken from the Buss and Plomin
This would tend to result in smaller correlations between attractiveness
Emotionality Activity Sociability Impulsivity temperament rating system
actors who were very high and very low on attractiveness. However, the
(EASI; 1975), and the Cronbach alphas for these scales when applied by
our judges ranged from .82 to .65.
use of extreme groups to test for the presence of relations generally results
in overestimation of the strength of relations (cf. Feldt, 1961).
and other ratings than would be obtained if we had deliberately selected
ATTRIBUTIONS OF PERSONALITY
795
indicated that there were no significant differences in the ratings given
by male and female judges, therefore sex was not included as a factor in
the analyses.
For the Actor X Information Type interaction, A = .634, ap-
Procedure
effects. However, both the main effects were also significant. For
For the overall MANOVA, all three of the effects were significant.
proximate F(60, 1886) = 2.860, p < .001. This substantial (and
disordinal) interaction makes it difficult to interpret the main
A set of 6 female actors was chosen randomly from the same pool of
26 actors that was used in the instrument development study. There were
18 conditions: 6 actors each contributed three types of information (photograph, speech sample, writing sample). In each condition, judges rated
only one of the actor by stimulus combinations in order to avoid possible
carryover effects. In the slide condition, a group of judges was shown a
color slide of one actor and asked to fill out the rating questionnaire. The
slide was displayed during the entire rating period. In the tape condition,
the tape (which was about 1 min in duration) was played once before the
judges began rating, and played a second time about 5 min into the rating
period. In the handwriting condition, each judge was given a photocopy
of one actor's handwriting and asked to rate the writer. Judges in both
the tape and handwriting conditions were informed that the content of
the message (a paragraph from an art history textbook) was provided to
the actor, and were instructed to ignore content and focus on the style of
the speech or writing.
the differences in the ratings received by the 6 actors, A = .833,
approximate f\W, 1438) = 2.25, p < .001; for the differences
in the ratings given to the three different types of information,
A = .0206, approximate F(\2, 10) = 4.971, p = .008. (This
extraordinarily small lambda was obtained when the interaction
term was used as the error term; if the within-cell error term is
used, A = .463, approximate^ 12, 718) = 28.12,;) < .001. The
main effect for information is significant using either error term.)
Interpreting the significant interaction in light of the hypotheses, it appears the pattern of judgments differs from one
information type to another. To obtain more information about
possible differences across-information types, a follow-up analysis
of the simple main effects of the actor was done separately for
each of the three information conditions. This included a discriminant function analysis to evaluate the relative contribution
of each of the six dependent variables to the discrimination
Results
among actors.
The results of the discriminant analyses that were done sep-
After preliminary data reduction through the formation of
arately for each information type are shown in Table 3. For all
summated scales (as reported in the Appendix), there were six
three information types, the first two dimensions of the five di-
dependent variables. Three were person perception variables de-
mensional solution are shown. Standard procedures for testing
rived from our own bipolar adjective ratings (social evaluation,
the significance of these roots were used (Nie & Hull, 1981). For
intellectual evaluation, and potency). Three were Buss temper-
all three of these information types, the set of Roots 1-5 was
ament scales (Emotionality, Activity, and Sociability). These
significant (p < .05). The set containing Roots 2-5 was significant
scales are not orthogonal; correlations between these scales ranged
(p = .033) for the slide condition, and fell just short of conven-
from —.383 to +.688. Because there were six correlated depen-
tional significance levels for the handwriting condition (p = .083)
dent variables, MANOVA was used to evaluate the predictive use-
and the tape recorded speech condition (p = .064). To simplify
fulness of these six personality dimensions.
interpretation of these results, the following rule was adopted: A
particular dependent variable was viewed as being a useful source
Multivariate Analysis of Variance
of discriminating information only if its beta weight for the standardized canonical discriminating function and its structure
A 6 X 3 MANOVA was performed using the six personality
coefficient (correlation with the discriminating function) both
scales (social evaluation, intellectual evaluation, potency, emo-
exceeded .40 in absolute value and had the same sign. In addition,
tionality, activity, and sociability) as dependent variables. The
the univariate F ratio for each variable was also examined to see
factors were the six levels of the actor identity factor and the
whether it was consistent with the beta weight and structure coef-
three levels of information type (photograph, speech, and hand-
ficient. Finally, only those dimensions that were significant in
writing). Thus, each of the 18 cells in the design corresponded
the dimension reduction analysis will be interpreted. Conse-
to an actor/information stimulus combination (e.g., Actor 1's
quently, Dimensions 1 and 2 can be interpreted for the slide
speech, or Actor 4's handwriting). Type of Information was
condition, but only Dimension 1 will be interpreted for the
treated as a fixed factor, and Identity of Actor was treated as a
handwriting and tape conditions. For the scales that met these
random factor, thus, the significance test for the main effect of
conditions the standardized beta weights and structure coeffi-
information type had to use the interaction component of the
cients are underscored in Table 3.
model as the error term. Because of unequal within-cell №, the
For the slide condition, the first dimension along which the
computational procedures for nonorthogonal analysis of variance
actors were discriminated by the judges seems to be best char-
(ANOVA) with regression-type partitioning of sums of squares
acterized in terms of social evaluation, although there were other
were used. Each effect was tested controlling for all other effects.
scales that had moderately strong correlations with this dimen-
There did not appear to be serious violations of the assumptions
sion. The social evaluation scale involves attributions of warmth,
of multivariate normality. Each individual dependent variable
friendliness, optimism, cheerfulness, and attractiveness. There
had an approximately normal distribution, and the within-cell
was a second orthogonal dimension along which the judges dif-
correlation matrix was reasonably homogeneous across-treatment
ferentiated the slides; the scales which had the highest correlations
conditions. The pooled within-cell correlation matrix among the
with this second dimension were intellectual evaluation and ac-
six dependent variables is shown in Table 2. Generally the cor-
tivity. Taken together, these two scales suggest a pattern of com-
relations were low, except that the Buss sociability scale correlated
petence and "quickness," possibly both mental and physical
.503 with the social evaluation scale and .541 with the Buss ac-
quickness. For the handwriting condition, it was clear that only
tivity scale.
one of the six scales was a useful discriminator among the actors:
796
REBECCA M. WARNER AND DAVID B. SUGARMAN
Table 2
Within-Cell Correlation Matrix of the Six Scales
Scale
Scale
1.
2.
3.
4.
5.
6.
Social Evaluation
Intellectual Evaluation
Potency
Activity
Sociability
Emotionality
Mean
.732"
.242
.139
.371
.503
-.160
.604
.265
.203
.005
-.107
.785
.378
.222
-.330
.866
.541
-.061
.911
-.274
2.36
2.46
1.89
2.09
2.35
.655
1.78
Note. Both means and standard deviations are simple univariate results.
* Standard deviations are on the diagonal.
the potency scale. For the tape recorded speech condition, the
basis on which the judges apparently differentiated the six tape
recorded speech samples was primarily activity and social evaluation.
A few other inferences can be tentatively drawn. For instance,
although the sociability scale had a significant univariate F ratio
for the differences among actors in the slide condition, it appears
that this significant result may have occurred because sociability
is correlated with two other scales that were more effective predictors (the social evaluation and activity scales). The sociability
scale could be seen as redundant with these other two scales, and
it appears to have been less effective in describing perceived differences among the actors. In general, the results of the discrim-
Table 3
Simple Effects
inant analysis were fairly consistent with the univariate F ratios
that are also reported in Table 3. The only variables that appeared
less useful within the context of the multivariate model than
might have been expected from the univariate F ratios were sociability (in the slide condition) and intellectual evaluation (in
the tape condition). Another consistent finding was that the
emotionality scale was not a useful discriminator in any of the
information conditions. This reflects the generally lower level of
consensus among judges in their preceptions of emotionality.
A large beta weight, structure coefficient, and univariate F
ratio for any trait implies that disagreements among judges were
relatively small, and that actors received consistently different
ratings on that trait variable. Thus, when a scale such as social
Analysis of Actors Within Each Level of the Information Factor
Dimension 2
Dimension 1
Scale
B
B
re
Univariate
re
F
-.273
-.672
.296
-.454
-.047
-.049
8.89
3.03
2.10
4.66
4.87
.45
<.001
.011
.065
<.001
<.001
.810
-.240
-.558
.192
-.091
.514
-.009
1.19
2.04
3.71
1.81
1.50
.314
.073
.003
.110
.190
.928
-.103
.792
-.253
-.141
-.452
.203
3.94
3.13
P
Slide condition
Social Evaluation
Intellectual Evaluation
Potency
Activity
Sociability
Emotionality
-.816
.348
-.256
-.177
-.112
-.029
-.885
.050
-.359
-.564
-.665
.190
-.040
-.707
.743
-.686
.219
.132
Handwriting condition
Social Evaluation
Intellectual Evaluation
Potency
Activity
Sociability
Emotionality
.312
-.222
-.887
-.165
.199
-.363
.232
-.376
-.801
-.299
.169
-.141
-.469
-.429
.429
-.524
1.01 1
.257
.27
Tape condition
Social Evaluation
Intellectual Evaluation
Potency
Activity
Sociability
Emotionality
.460
-.177
-.426
1.069
-.498
.022
.500
.034
-.123
.771
.210
.179
-.120
.929
-.408
.041
-.310
.066
.53
7.92
1.99
.75
.002
.009
.755
<.001
.079
.585
Note. B = Standardized beta coefficient; re = Canonical structure coefficient. Underscored numbers denote dependent variables with both beta and
structure coefficient greater than .40, only for significant dimensions.
ATTRIBUTIONS
evaluation is a useful predictor within the discriminant analysis,
this suggests that judges showed better agreement in their assignment of ratings of social evaluation than in their assignment
of ratings of other variables such as emotionality. An alternative
method of assessing agreement among judges would be to treat
each scale separately in a univariate ANOVA and use the intraclass
correlation as an index of reliability of the assessments made by
each individual judge. This univariate analysis provides a picture
of the relative importance of variables that is generally similar
to the picture that is obtained from the multivariate analysis,
apart from the exceptions noted earlier.
Discussion
Two hypotheses were discussed in the Introduction. This study
does not provide strong support for the source consistency hypothesis; the significant and disordinal interaction between information type and actor makes it difficult to interpret the significant differences in the ratings received by actors. This study
provides clearer support for the differential information hypothesis, which stated that each information type or communication
channel provides information about different personality dimensions. Airport's statement (1961) that there would be some
redundance across channels is not being disputed here. However,
these results suggest that each communication channel (facial
appearance, speech, handwriting) provides information about a
different collection of personality dimensions, and that a particular attribute (such as potency) can sometimes be judged more
consistently from one channel (such as handwriting) than from
other channels. Specifically, actors were differentiated along a
potency dimension much more clearly based on the handwriting
samples than based on speech or facial appearance. Actors were
differentiated along an activity dimension more consistently based
on their speech than based on other communication channels.
For other personality attributes such as social and intellectual
evaluation, judges were able to make the most consistent differentiation among actors based on the facial appearances.
The results obtained here are basically in agreement with earlier research. For instance, the handwriting research just cited
has consistently found correlations between "potency" type
variables (status, self-esteem, etc.) and various features of handwriting such as signature size, although some authors feel that
this correlation may exist only for female writers and not for
males (Jorgenson, 1977). In this study, when judges were given
the opportunity to make attributions using six personality scales,
potency was the only scale that yielded consensus among the
judges of handwriting samples.
Another point of agreement between this study and earlier
findings is the importance of the evaluative dimensions when
judging physical appearance. Existing research on physical attractiveness suggests that "what is beautiful is good" (Dion et
al., 1972). Judges invoke a generally positive personality stereotype when they make attributions about the personality or abilities
of physically attractive persons. In the present study, both of the
evaluative scales (Social and intellectual evaluation) showed
greater consensus among judges of slides than judges of tapes or
handwriting. However, the slides also elicited fairly consistent
attributions about activity and sociability characteristics that were
only moderately correlated with the evaluative scales.
OF
797
PERSONALITY
Existing research did not point to a simple description of attributions based on tape recorded speech; in the present study,
the judges of tapes showed the highest level of agreement when
judging activity, and lower levels of agreement when judging social
evaluation, sociability, and intellectual evaluation. Given earlier
research linking speech to judgments of anxiety or emotionality,
the inability of the judges to agree on the emotionality ratings
of tape recorded speech in this study seems anomalous. One
potential explanation is that actors read an emotionally neutral
paragraph from a textbook probably limited the amount of information about emotionality that was available from speech.
There was wide variability in the emotionality ratings given, but
judges did not agree which voices belonged to more or less emotional speakers. The lack of consensus among judges in rating
emotionality from speech in this study may represent a methodological artifact rather than a general lack of emotional information in speech.
This study has several limitations that restrict the types of
conclusions that can be drawn. First, there were only 6 female
actors. The results may be partly due to peculiarities of these
particular actors, and certainly cannot be generalized to males
until the findings have been replicated using male actors. Second,
the selection of adjectives and rating scales does not cover all
possible personality dimensions. The Osgood, Suci, and Tannenbaum (1957) three dimensional semantic differential (evaluation, activity, potency) guided the selection of scales, because
numerous studies in social psychology and nonverbal communication have found some variant of these dimensions to be useful
(e.g., Brown et al., 1974; Exline, 1972; Hayes & Meltzer, 1972;
Scherer, 1974; Wish, 1978). This was supplemented with the
Buss and Plomin (1975) temperament scales. However, there may
well be other attributions that judges could make reliably based
on these kinds of information that were not represented in the
selection of scales used here. Third, the information types could
have been defined differently, as in studies that distinguish between transcripts of verbal content and content-filtered vocal
characteristics. The intent in designing this study was to use
relatively naturalistic expressive behaviors; rather than asking
actors to "act out" specific emotions or attempt to convey a
"warm" or "cold" impression, the stimuli were obtained by recording the actors' spontaneous reactions to the request to speak,
write, and pose for a photograph. Control over the specific information content of the stimuli was sacrificed to obtain some
improvement in the "ecological validity" (Brunswik, 1956) of
the social stimuli used as the basis for the judges' attributions.
The results of this study corroborate earlier findings of differential information about personality contained in facial appearance, handwriting, and speech (cf. Burns & Beier, 1973; Ekman
etal., 1980; and Zuckerman et al., 1982). The results imply that,
rather than attempting to show how much weight is given to each
information channel in general, it may be necessary to ask how
much weight is given to each information channel for each specific attribute that is being judged.
References
Adams, G. R. (1977). Physical attractiveness research: Toward a developmental social psychology of beauty. Human Development, 20, 217-
238.
798
REBECCA M. WARNER AND DAVID B. SUGARMAN
Aiken, L. R., & Zweigenhaft, R. L. (1978). Signature size, sex, and status
in Iran. Journal of Social Psychology, 106, 273-274.
Feldt, L. S. (1961). The use of extreme groups to test for the presence of
a relationship. Psychometrika, 26, 307-316.
Allport, G. W. (1961). Pattern and growth in personality. New York: Holt,
Rinehart & Winston.
Allport, G. W, & Cantril, H. (1934). Judging personality from voice.
Journal of Social Psychology, 5, 37-55.
Allport, G. W, & Vernon, P. F. (1933). Studies in expressive movement.
New York: MacMillan.
Baron, R. M. (1981). Social knowing from an ecological-event perspective:
A consideration of the relative domains of power for cognitive and
perceptual modes of knowing. In J. H. Harvey (Ed.), Cognition, social
behavior, and the environment. Hillsdale, NJ: Erlbaum.
Herman, H. J., Shulman, A. D., & Marwit, S. J. (1976). Comparison of
multidimensional decoding of affect from audio, video and audiovideo
recordings. Sociometry, 39, 83-89.
Berscheid, E., & Walster, E. (1974). Physical attractiveness. In L. Berkowitz
(Ed.), Advances in experimental social psychology (Vol. 7, pp. 157215). New York: Academic Press.
Brown, B. L., Strong, W. L., & Rencher, A. C. (1974). Fifty-four voices
from two: The effects of simultaneous manipulations of rate, mean
Guise, B. J., Pollans, C. H., & Turkat, I. D. (1982). Effects of physical
attractiveness on perception of social skills. Perceptual & Motor Skills,
54, 1039-1042.
Hall, C. S., & Lindzey, G. (1970). Theories of personality (2nd ed.). New
York: Wiley.
Hayes, D. P., & Meltzer, L. (1972). Interpersonal judgments based on
talkativeness. 1: Fact or artifact? Sociometry, 35, 538-561.
Jones, W. H., Hannson, R. C., & Phillips, A. L. (1978). Physical attractiveness and judgments of psychopathology. Journal of Social Psychology, 105, 79-84.
Jorgenson, D. O. (1977). Signature size and dominance: A brief note.
Journal of Psychology, 97, 269-270.
Krauss, R. M., Apple, W., Morency, N., Wenzel, C, & Winton, W. (1981).
Verbal, vocal, and visible factors in judgments of another's affect. Journal of Personality and Social Psychology, 40, 312-320.
Mehrabian, A., & Ferris, S. R. (1967). Inference of attitudes from nonverbal communication in two channels. Journal of Consulting Psychology, 31, 248-252.
fundamental frequency, and variance of fundamental frequency on
Nie, C. H., & Hull, N. H. (1981). SPSS Update 7-9. New York: McGraw-
ratings of personality from speech. Journal of the Acoustical Society
of America, 55, 313-318.
Hill.
Osgood, C. E., Suci, G. J., & Tannenbaum, P. H. (1957). The measurement
Brunswik, E. (1956). Perception and the representative design of psycho-
of meaning. Urbana: University of Illinois Press.
logical experiments. Berkeley: University of California Press.
Bull, R., & Stevens, J. (1979). The effects of attractiveness of writer and
Ostwald, P. F. (1965). Acoustic methods in psychiatry. Scientific American,
penmanship on essay grades. Journal of Occupational Psychology, 52,
53-59.
Rosenthal, R., Hall, J. A., DiMatteo, M. R., Rogers, P. L., & Archer, D.
(1979). Sensitivity to nonverbal communication: The PONS test. Bal-
Burns, K. L., & Beier, E. G. (1973). Significance of vocal and visual
212. 82-91.
timore: Johns Hopkins University Press.
channels in the decoding of emotional meaning. Journal of Communication, 23, 118-130.
Scherer, K. R. (1974). Acoustic concomitants of emotional dimensions:
Buss, A., & Plomin, R. (1975). A temperament theory of personality
verbal communication: Readings with commentary. New York: Oxford
Judging affect from synthesized tone sequences. In S. Weitz (Ed.), NonUniversity Press.
development. New York: Wiley.
Cronbach, L. J. (1955). Processes affecting scores on "understanding of
Scherer, K. R., & Giles, H. (1979). Social markers in speech. Cambridge:
others" and "assumed similarity". Psychological Bulletin, 52, 177193.
Cambridge University Press.
Unger, R. K., Hilderbrand, M., & Madar, T. (1982). Physical attractiveness
DePaulo, B. M., Rosenthal, R., Eisenstat, R. A., Rogers, P. L., & Fm-
and assumptions about social deviance: Some sex-by-sex comparisons.
Personality and Social Psychology Bulletin, 8, 293-301.
kelstein, S. (1978). Decoding discrepant nonverbal cues. Journal of
Personality and Social Psychology, 36, 313-323.
Dion, K., Berscheid, E., & Walster, E. (1972). What is beautiful is good.
Williams, C. E., & Stevens, K. N. (1972). Emotions and speech: Some
Journal of Personality and Social Psychology, 24, 285-290.
Edinger, J. A., & Patterson, M. L. (1983). Nonverbal involvement and
1238-1250.
Wish, M. (1978). Dimensions of dyadic communication. In S. Weitz
acoustical correlates. Journal of the Acoustical Society of America, 52,
(Ed.), Nonverbal communication: Readings with commentary (2nd ed.).
social control. Psychological Bulletin. 93, 30-56.
Ekman, P., Friesen, W. V., O'Sullivan, M., & Scherer, K. (1980). Relative
affect. Journal of Personality and Social Psychology, 38, 270-277.
New York: Oxford University Press.
Zuckerman, M., Amidon, M., Bishop, S., & Pomerantz, S. (1982). Face
and tone of voice in the communication of deception. Journal of Per-
Exline, R. V. (1972). Visual interaction: The glances of power and preference. In J. Cole (Ed.), Nebraska Symposium on Motivation, 1971.
Zweigenhaft, R. L. (1977). The empirical study of signature size. Social
importance of face, body, and speech in judgments of personality and
Lincoln: University of Nebraska Press.
sonality and Social Personality, 43, 347-357.
Behavior and Personality, 5, 177-185.
(Appendix follows on next page)
ATTRIBUTIONS OF PERSONALITY
799
Appendix
Rating scales were developed in a preliminary instrument development
study, using a separate group of 273 judges to provide data on factor
Scale scores were created using unweighted linear composites for each
of these three groups of variables. Factor score coefficients could have
structure and internal consistency. For the initial item pool, 32 pairs of
been used to create factor scores; however, simple unweighted linear
bipolar adjectives were generated. Eight adjective pairs were chosen by
composites were chosen for several reasons including simplicity; consis-
the principal investigators and their research assistants for each of the
tency of scoring of the Buss scale and the new scales created for this
Mowing four domains: activity, potency, social evaluation, and intellectual
evaluation. Face validity was the criterion for inclusion in this initial
to describe the internal homogeneity for scales that are simple unweighted
phase. A questionnaire was set up incorporating these 32 bipolar adjective
linear composites. Cronbach alphas were calculated to assess the internal
pairs with a 5-point rating scale. Ratings were to be made relative to the
homogeneity of the scales created by averaging the scores for the groups
"average college student." Thus, if the bipolar adjective pair quiet/noisy
were rated on a 5-point Likert scale: about as noisy as an average college
of items. The social evaluation scale alpha was .773; the intellectual evaluation alpha was .833; and the potency scale alpha was .747. Correlations
study; and the availability of simple statistics such as Cronbach's alpha
student (3); a little bit quieter than average (2); much noisier than average
among these three scales tended to be small; the largest correlation was
(5). The position of the more socially desirable adjective was varied ran-
r = .354 for the intellectual and social evaluation scales. These three
domly so that sometimes the left side was the more socially desirable and
sometimes the right side was the more socially desirable; however, to
scales seemed to summarize the major dimensions of person perception
that were implicit in our original selection of 32 adjectives reasonably
well.
facilitate scoring and analysis, the data were receded with 5 always corresponding to the more positive end of the scale. The questionnaire also
After the results of the main study were obtained, the ratings of the
asked for the sex, age and year in college of the rater, and asked about
previous familiarity with the actor.
382 judges were factor analyzed to evaluate the stability of the factor
structure. The rotated factor loadings are reported in Table 1. The results
A group of 273 judges each rated one actor/information combination
using the 32 bipolar adjectives. The data were pooled across the 18 con-
were essentially identical to those in the instrument development phase
ditions and factor analyzed. This procedure involved the implicit assumption that there was homogenous factor structure across the 18 con-
prior to rotation, and the order of the factors was different. Alpha coefficients were calculated to assess the internal homogeneity of the three
except that the first three factors only accounted for 55% of the variance
ditions, an assumption that was not tested directly because the number
scales, which were composed of the same items that were used previously.
of subjects within each cell was too small to obtain reliable estimates of
There was some shrinkage of the alphas, but the scales were still sufficiently
within-cell correlations and structure. The initial principal factors analysis
reliable to be used: the social evaluation scale alpha was .776; the intel-
was conducted with varimax rotation. All 32 bipolar adjectives were in-
lectual evaluation scale alpha was .759; and the potency score alpha
cluded, and there were seven factors with eigenvalues greater than one.
was .706.
Analysis was also done to check the internal consistency of the items
in the four Buss EASI scales within our sample. Alphas were .65 for the
Examination of these factors indicated that the last four factors consisted
of single items or pairs of items. The set of eight adjectives that were
supposed to detect an "activity" dimension did not appear together on
a factor; instead, these items tended to load moderately highly on the
factors which seemed primarily interpretable as intellectual evaluation,
emotionality scale, .79 for the sociability scale, .82 for the activity scale,
and .45 for the impulsivity scale. Due to the low alpha value, the impulsivity scale was not included in subsequent analyses. Also, the item
social evaluation, and potency. Based on this initial analysis it was decided
to retain five items as measures of intellectual evaluation; five items as
"is independent of others" was omitted from the Buss sociability scale,
measures of social evaluation; and three items as measures of potency.
A second factor analysis was performed using this reduced set of 13
men and a positive correlation for women (Buss & Plomin, 1975, p. 26).
because it had a negative correlation with other items on this scale for
items; the rotated factor loadings are shown in Table 1. Factor 1 was
labeled an Intellectual Evaluation factor; Factor 2 as a Social Evaluation
factor; and Factor 3 was named a Potency factor. Before rotation, the
first three factors accounted for 60% of the variance, and only these three
factors had eigenvalues greater than one.
Received April 4, 1985
Revision received August 23, 1985