Abstract
The self-serving bias is the tendency to consider oneself in unrealistically positive terms. This phenomenon has been documented for body attractiveness, but it remains unclear to what extent it can also emerge for own body size perception. In the present study, we examined this issue in healthy young adults (45 females and 40 males), using two body size estimation (BSE) measures and taking into account inter-individual differences in eating disorder risk. Participants observed pictures of avatars, built from whole body photos of themselves or an unknown other matched for gender. Avatars were parametrically distorted along the thinness–heaviness dimension, and individualised by adding the head of the self or the other. In the first BSE task, participants indicated in each trial whether the seen avatar was thinner or fatter than themselves (or the other). In the second BSE task, participants chose the best representative body size for self and other from a set of avatars. Greater underestimation for self than other body size emerged in both tasks, comparably for women and men. Thinner bodies were also judged as more attractive, in line with standard of beauty in modern western society. Notably, this self-serving bias in BSE was stronger in people with low eating disorder risk. In sum, positive attitudes towards the self can extend to body size estimation in young adults, making own body size closer to the ideal body. We propose that this bias could play an adaptive role in preserving a positive body image.
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Notes
In the psychometric estimates task, BSE blocks for self and other were counterbalanced and this meant that half of the participants in our study completed the judgment on the self after the judgment on the other. Hence, exposure to the other person’s body could, in principle, have influenced the subsequent rating on the self. To assess whether this influence occurred we ran explorative analyses that considered block order (self-first vs. other-first) as a factor. Specifically, we repeated the ANCOVAs described in the Analyses section adding block order (self-first vs. other-first) as a between-participant factor. None of these analyses revealed a significant main effect or interaction involving block order (all ps > 0.17). It should be remarked that the other person shown in the video was selected to be in the normal BMI range and of average attractiveness. Hence, unlike previous works (e.g., Cazzato et al., 2016) we did not use thin or overweight others that could trigger comparison judgments. Furthermore, it is important to note that our methods of constant stimuli entailed the presentation of an equal number of thinner/fatter distortions of the other. This could have also contributed to cancel out any potential bias related to seeing the other’s body before judging the self. Because our participants could differ in BMI with respect to the neutral other, we also repeated the analyses reported above, replacing EDR, with the participant’s BMI as a covariate. For PSE and body size attractiveness, none of these analyses revealed a significant main effect or interaction involving the block order variable (all ps > 0.11). For the Direct choice task, a main effect of block order (F(1, 69) = 5.91, p = 0.02, \(\eta _{{\text{p}}}^{2}\) = 0.08) and the interaction between block order and BMI (F(1, 69) = 7.10, p = 0.01, \(\eta _{{\text{p}}}^{2}\) = 0.09) reached significance. Critically, however, these effects of block order did not involve the Target variable, ps > 0.17. All other effect involving block order, ps > 0.14.
In our experimental design we counterbalanced the experimenter’s gender across participants (see Procedure section). To ascertain any potential influence of this factor on our dependent variables we also repeated the ANCOVAs described in the Analyses section adding Experimenter’s Gender as a between-participants factor, separately for each of our main dependent variables (BSE: PSE and direct choice; body size attractiveness). For the PSE we only found a main effect of Experimenter’s Gender, F(1, 69) = 5.65, p = 0.02, \(\eta _{{\text{p}}}^{2}\) = 0.08. Participants tested by the male experimenter generally underestimated more (M = − 7.40%, SD = 5.53) compared to those tested by the female experimenter (M = − 4.12%, SD = 4.64). Importantly, however, the Experimenter’s Gender did not interact with the other factors (all p > .10), indicating that this counterbalanced variable did not influence the SSB in male or female participants (i.e., there was no thee way interaction between Experimenter’s Gender, Participants’ Gender and Target; p = 0.12). For the Direct choice task, all main effects or interactions involving Experimenters’ Gender were not significant (all ps > 0.24, except for a marginal main effect of Experimenters’ Gender, F(1, 69) = 3.35, p = 0.07, \(\eta _{{\text{p}}}^{2}\) = 0.05). For body size attractiveness, we found a main effect of Experimenter’s Gender, F(1, 69) = 4.71, p = 0.03, \(\eta _{{\text{p}}}^{2}\) = 0.06. Participants tested by the male experimenter generally preferred thinner body distortions (M = − 13.40%, SD = 6.82) compared to those tested by the female experimenter (M = − 12.43%, SD = 7.16). The interaction between the Experimenters’ Gender interaction and EDR, F(1, 69) = 5.71, p = 0.02, \(\eta _{{\text{p}}}^{2}\) = 0.08, suggested that preference for thinner body distortions increased as EDR increases in those tested by the female experimenter, b = − 0.27, β = − 0.38, p = 0.02, but not in those tested by the male experimenter, b = 0.18, β = 0.31, p = 0.06. All other ps > .22 (except for a marginal interaction effect between Target and Experimenters’ Gender, F(1, 69) = 3.45, p = 0.07, \(\eta _{{\text{p}}}^{2}\) = 0.05).
After the direct choice task was performed on self and other, a direct body estimate was also repeated for the experimenter’s body. In this case, the Target factor included three levels: self, other and experimenter. The main effect of Target, F(2,138) = 8.01, p < 0.001, \(\eta _{{\text{p}}}^{2}\) = 0.10, showed that both experimenters (M = − 13.56%, SD = 9.89) were underestimated more than the self (M = − 5.96%, SD = 9.71) or the other (M = − 5.66%, SD = 7.58), ps < .001. Target interacted with EDR, F(2,138) = 4.71, p = 0.01, \(\eta _{{\text{p}}}^{2}\) = 0.06. EDR predicted body size estimation for the self (linear regression: b = 0.25, β = 0.28, p = 0.01; Figure 3d) but not for the other (linear regression: b = −.12, β = − 0.17, p = 0.14; Figure 3e) nor for the experimenters (linear regression: b = 0.001, β = 0.002, p = 0.99). Finally, Target interacted also with participants’ Gender, F(2,138) = 3.89, p = 0.02, \(\eta _{{\text{p}}}^{2}\) = 0.05. Women and men did not give different judgement for the self (women: M = − 7.03%, SD = 10.63; men: M = − 4.92%, SD = 8.73), p = 0.35, and the experimenter (women: M = − 12.76%, SD = 10.52; men: M = − 14.33%, SD = 9.31), p = 0.49, while women underestimated more the other seen in the video (M = − 7.68%, SD = 7.11) compare to men (M = − 3.69%, SD = 7.59), p = 0.02. Finally, both experimenters were judged thinner than the self and the other, ps < 0.02. All other ps > 0.09.
The correlation between the difference in BSE for self and other (i.e., our measure of SSB) and eating disorder risk remained significant even when controlling for Self-objectification (psychometric estimates: r = 0.44, p < 0.001; direct choice: r = 0.39, p = 0.001), Body surveillance (psychometric estimates: r = 0.45, p < 0.001; direct choice: r = 0.36, p = 0.001), Body shame (psychometric estimates: r = 0.42, p < 0.001; direct choice: r = 0.38, p = 0.001) and PACS (psychometric estimates: r = 0.47, p < 0.001; direct choice: r = 0.39, p < 0.001).
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Acknowledgements
This study was funded by Fondazione Cassa di Risparmio di Trento e Rovereto (IT), Bando giovani ricercatori 2014 to M.M. We would like to thank Fabio La Russa who helped with data collection, Andrea Zignoli and Anna Colombatto who acted “the other” in the videos.
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Mazzurega, M., Marisa, J., Zampini, M. et al. Thinner than yourself: self-serving bias in body size estimation. Psychological Research 84, 932–949 (2020). https://doi.org/10.1007/s00426-018-1119-z
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DOI: https://doi.org/10.1007/s00426-018-1119-z