The Politics of Being “Cait”: Caitlyn Jenner,Transphobia,and Parasocial Contact Effects on Transgender-Related Political Attitudes

Abstract

Celebrities saturate American culture and often become relevant in politics, yet political science has largely left unstudied how celebrities affect mass political behavior. We focus on the 2015 story of Caitlyn Jenner revealing her transgender identity. Using an original nationally representative survey from that summer, we examine whether following the Jenner story and evaluations of its social significance affected attitudes toward transgender rights policies. Specifically, we examine how age and transphobia interacted with engagement with the Jenner story to shape attitudes toward transgender rights. We find, counterintuitively, that older respondents who were more transphobic were less likely to see her story as representing negative social trends if they followed it in the media. Furthermore, more transphobic older respondents were more likely to support pro-transgender policies if they viewed Jenner’s story less negatively. We then discuss the implications of our findings for research on celebrity effects on politics and transgender rights.

Keywords

celebrities Caitlyn Jenner parasocial contact transgender transgender rights transphobia

Caitlyn Jenner is one of the best known openly transgender celebrities in the United States (Lovelock, 2017). By 2015, when she publicly confirmed that she was a transgender woman, Jenner had enjoyed over 40 years of prominence as an athlete, product spokesperson, and television star. This disclosure was a high-profile story when, after significant tabloid speculation, Jenner discussed her gender identity with journalist Diane Sawyer, and appeared on the cover of Vanity Fair under the headline “Call me Caitlyn” (Bissinger, 2015).

Celebrities like Jenner are important in politics, both in campaigns and issue advocacy (Harvey, 2017; West & Orman, 2003). However, Jenner’s political relevance differs from more common forms of celebrity activism in that her mere act of “coming out” could be seen as political. Identifying as a member of a marginalized community increases its visibility, and indeed the term transgender itself “refers to a collective political identity” (Currah et al., 2006, p. xv). Her revelation coincided with rising attention to transgender rights, including several local referendums that year on nondiscrimination ordinances that received substantial national media coverage (Taylor et al., 2018). By coming out in that environment, coupled with the fact that relatively few Americans report knowing a transgender person (Tadlock et al., 2017), Jenner was positioned to be a familiar face representing a group whose rights were increasingly salient but with whom most Americans have limited personal experience.

In this research, we treat Jenner’s story as a case of parasocial contact—individuals forming “beliefs and attitudes about people they know only through” mass media (Schiappa et al., 2006, p. 20)—that may influence group-relevant attitudes. Based on extant literature, we focus on the direct and conditional effects of age and transphobia in predicting perceptions of the Jenner story and how those perceptions affected transgender rights policy attitudes. Counterintuitively, we find that older respondents who were also more transphobic were less likely to see her story as representing negative social trends if they followed the story more closely and were more likely to translate that evaluation into support for pro-transgender policies.

Celebrity and Politics

A “celebrity” is someone “who has a high public profile, usually promoted by appearances in the mass media, and they are consequently recognized by others” (Gunter, 2014, p. 2). That fame may derive from factors such as personality, occupation, lifestyle, or attractiveness (Gunter, 2014; Marwick, 2013). Critically, though politicians may gain national fame through public affairs, the academic literature differentiates them from celebrities because celebrities usually gain fame through other less overtly political means and then may adopt a political career or platform once that fame is established (Harvey, 2017).

Celebrities play many political roles. They champion issues, endorse and fundraise for candidates, or even run for office themselves (Harvey, 2017; West & Orman, 2003). Although celebrity endorsements are not a new phenomenon, scholars have only recently explored how those endorsements shape citizen behavior. Research on the American, Canadian, and British contexts suggests that celebrity endorsements can influence voter turnout, issue preferences, issue salience, candidate support, perceptions of candidate traits or viability, and voter learning even when persuasion does not occur (Austin et al., 2008; Chou, 2015; Inthorn & Street, 2011; Jackson, 2008; Jackson & Darrow, 2005; McClerking et al., 2019; Nownes, 2012; Pease & Brewer, 2008; Street et al., 2008).

One trend in celebrity endorsement studies, especially in consumer and health behavior research, is that celebrities tend to have stronger influence on youth than on older persons (Austin et al., 2008; Buksa & Mitsis, 2011; Dix et al., 2010; Inthorn & Street, 2011; Jackson & Darrow, 2005; LaFerle & Chan, 2008; Stevens et al., 2003). These studies find that younger people are more interested in and attentive to celebrity culture, and that the youth audience is a larger target both for entertainment media and brands seeking celebrity endorsements. A common thread across these studies is that youth are at a more impressionable stage of cognitive development that makes them more susceptible to cues from celebrities who they follow, regard positively, identify with through shared identities, or who are popular among their peers.

Jenner’s fame, though, is multigenerational, and may defy the trend of greater celebrity influence on youth given how successive generations have known her in different contexts. She gained notability as a man, winning a gold medal in the decathlon at the 1976 Summer Olympics. Jenner parlayed this into major endorsements, most notably for Wheaties cereal. She remained visible with more than 100 film and television appearances—often portraying herself—that included long-term stints on highly rated shows (IMDb, 2019). From 2007 to 2017, Jenner was a major character on the popular reality show Keeping Up with the Kardashians.

After coming out, Jenner starred in her own reality show I Am Cait that focused on her gender transition. Controversially, she received the Arthur Ashe Courage Award in 2015 to honor her openness about her gender identity, and quickly gained millions of social media followers and praise from President Barack Obama for her coming out (Lovelock, 2017). After our survey was completed, Jenner made explicitly political statements about her conservative politics and support for Donald Trump, but also her pro-transgender policy positions (Howard, 2016).

Effects of Parasocial Contact

The Jenner story differs from most political science literature on celebrities because when Jenner revealed her transgender identity, she was neither a candidate nor was she endorsing a candidate. Rather, her political act was to be open about her identity at the time when transgender rights were becoming more salient. Thus, while research on celebrities may still be relevant for any effects that Jenner had on political behavior, the mechanism behind any such effects may differ from the typical dynamics of candidate choice or persuasive endorsement effects.

We posit that Jenner is a case of parasocial contact. As Harvey (2017) notes, typically “celebrities are not personally known to their audiences, yet they are connected through ‘parasocial interaction’, a phenomenon when people experience intimacy and familiarity with individuals and characters” through mass media (p. 18). The theory of parasocial contact extends interpersonal contact theory, which argues that contact between members of majority and minority groups can reduce majority group prejudice toward the minority group (see Schiappa et al., 2006 for more thorough review). Rothbart and John (1985) assert that attitudes about social groups are often formed in the absence of knowing members of those groups in more than a minimal fashion—if at all—but that interaction with outgroup members increases learning and familiarity that cause individuals to modify or elaborate on their prior attitudes toward a group, often positively.

Scholars argue that relationships formed with celebrities—including reporters, athletes, and reality show stars—or fictional characters through mass media can have the same effects as “real life” interactions despite differences in context and quality from in-person relations (Marwick, 2013; Schiappa et al., 2006). Research suggests that the brain processes parasocial interactions and direct interpersonal interactions similarly, causing comparable behavioral responses (Kanazawa, 2002; Reeves & Ness, 1996). Indeed, Perse and Rubin (1989) argue that the ability of television to induce parasocial interactions may make it more important to attitudes than real life interactions given the relative homogeneity and insularity of human social networks versus the greater diversity encountered on television.

Studies find reduced prejudice toward lesbian, gay, bisexual, and transgender (LGBT) populations resulting from exposure to LGBT persons or fictional LGBT storylines in mass media (Bond & Compton, 2015; Flores et al., 2018; Galinec & Korajlija, 2017; Garretson, 2015; Gillig et al., 2018; Hoffarth & Hodson, 2018; Jones et al., 2018; McDermott et al., 2018; Riggle et al., 1996; Schiappa et al., 2006; Solomon, Kurtz-Costes, 2018). Where transgender persons are concerned, this research finds ameliorating effects on various cognitive and affective measures of anti-transgender bias, but disagrees over whether parasocial interactions induce more pro-transgender policy preferences.

Especially critical for transgender persons, parasocial interaction effects tend to be stronger when direct contact with the group being portrayed in media is minimal (Armstrong & Neuendorf, 1992; Fujioka, 1999; Tan et al., 1997). Less than 1% of Americans identify as transgender, and few Americans report knowing a transgender person (Tadlock et al., 2017). For example, Tadlock et al. (2017) find that only 15% of Americans know a transgender person in any capacity—far fewer than the number that knows a gay or lesbian person. Importantly, past research does not find that knowing a transgender person is a consistent predictor of attitudes toward that group. Given the stigma surrounding transgender Americans (Lewis et al., 2017), they may be more comfortable revealing their identities to others who they suspect will accept them, meaning that personal contact may have little room to increase tolerance. Consequently, knowing a transgender person may have little to no statistical effect on transgender-related attitudes.

As a result, most Americans encountering the Jenner story via the mass media did not personally know a transgender person, especially those who transgender people may not come out to if they are perceived as potentially unaccepting. Jenner may be the first transgender person that many encountered either directly or parasocially. And, even among those who know a transgender person, Jenner could be the first familiar person they “knew” throughout their gender transition. Any parasocial effects associated with Jenner, then, may be especially strong given how isolated most of the public is from transgender persons and their identity narratives.

Transphobia as a Potential Influence

Individuals likely process the Jenner story through various predispositions and biases. Research shows that influences on American attitudes toward transgender rights include: affect toward transgender persons, authoritarianism, disgust sensitivity, core values, gender conformity, identity priming and gender roles, contact with gays and lesbians, elite cues, need for closure, and predictable demographic correlates (Flores, 2015; Flores et al., 2018; Haider-Markel et al., 2017; Harrison & Michelson, 2019; Jones & Brewer, 2018; Jones et al., 2018; Miller et al., 2017; Norton & Herek, 2013; Tadlock et al., 2017). Although any of these could affect engagement with the Jenner story, our focus is on negative bias toward transgender persons, termed “transphobia.”

Researchers often casually label any negative sentiments toward gender nonconforming people as “transphobia.” However, Hill and Willoughby (2005) propose a “Genderism and Transphobia scale” (GTS) to capture dislike and bias toward “individuals who do not conform to society’s gender expectations” (p. 533). To date, the GTS has not received attention in political behavior literature, but we believe that it may be relevant for the Jenner story in a way that known correlates of transgender-related attitudes might not. The GTS is multidimensional. Hill and Willoughby (2005) label one of the dimensions as “transphobia.” This dimension assesses biased attitudes, including items about physical gender transition through surgery or other means, how gender is presented via clothing, and the morality of gender self-presentation.

The transphobia dimension may especially resonate for Jenner’s story. As Olympian Bruce Jenner, her fame epitomized masculine ideals through her professional roles (Lovelock, 2017). However, her gender transition involved surgical alterations, a change in her outward presentation of gender, and a new name—changes that required Americans to refamiliarize themselves with a celebrity who many likely felt that they already knew well. These typical steps of gender transition are the core of what the transphobia dimension measures, so it may provide unique leverage over how Americans engage with Jenner in a way that other dispositions may not.

Hypotheses

In our analysis, we examine attention to and evaluations of the Jenner story. We further explore how engagement with her story relates to transgender rights policy attitudes. Given the previous discussion, our analytical focus is on age and transphobia. We propose the following hypotheses:

Hypothesis 1 (H1): Older persons will be less likely to follow the Jenner story than younger persons.

Hypothesis 2 (H2): More transphobic persons will be less likely to follow the Jenner story than less transphobic persons.

Hypothesis 3 (H3): Older persons will be less likely to evaluate the Jenner story positively than younger persons.

Hypothesis 4 (H4): More transphobic persons will be less likely to evaluate the Jenner story positively than less transphobic persons.

Hypothesis 5 (H5): More transphobic persons will hold less supportive attitudes toward transgender rights policies.

Hypothesis 6 (H6): Those evaluating the Jenner story more positively will have more supportive attitudes toward transgender rights policies.

Data

Our data are from an original survey of American adults that focused on transgender rights. It was funded by the University of Kansas, the University of Toledo, Ohio University, and the Williams Institute. Clear Voice Research (CVR) fielded the survey (N = 1,940) from June 12–25, 2015 using a web-based sample. CVR required double opt-in to surveys, IP verification, USPS address verification, and phone or SMS verification. Participants were recruited from existing CVR respondent pools via e-mail. Of 51,492 invites, the 1,940 completed responses yielded a participation rate of 3.78%, fairly common for commercial web-based surveys in academic research (Gideon, 2012). The survey demographics, reported in detail below, closely resembled the 2016 American National Election Study (ANES) time series (further information: www.electionstudies.org). This increased confidence in the representative nature of our survey and the validity of the analyses. Missing data were estimated via multiple imputation to avoid systematic bias from listwise deletion.

Measures

Following the Jenner Story

Respondents were asked, “Please tell me if you happened to follow each recent news story very closely, fairly closely, not too closely, or not at all closely.” Five stories prominent in the national media at the time were presented in randomized order, including “Bruce Jenner revealing that he is now Caitlyn Jenner.” Responses were coded 0 to 3 with higher scores representing closer attention (M = 1.36, SD = 0.99).

Evaluation of Jenner Story

Respondents answered, “Do you think the recent self-revelation that Bruce Jenner now identifies as Caitlyn Jenner is an isolated incident or a sign of broader changes in American society?” Response options included the following: “No opinion,” “An isolated incident,” “A sign of broader changes for the better,” and “A sign of broader changes for the worse.” Respondents lacking an opinion were dropped from the analysis (see Supplemental Appendix A for further information on “no opinion” responses to this item), and the remaining responses were arrayed as an ordinal variable scaled ±1 and ordered from lowest to highest value as: change for the worse, isolated incident, and change for the better (M = −0.03, SD = .84).

Transphobia

Hill and Willoughby (2005) proposed a 25-item transphobia scale. We asked five items that had factor loadings ≥0.70 on this dimension in their analysis, including: (a) “Sex change operations are morally wrong”; (b) “If a friend wanted to have his penis removed to become a woman, I would openly support him”; (c) “A man who dresses as a woman is a pervert”; (d) “It is morally wrong for a woman to present herself as a man in public”; and (e) “God made two sexes and two sexes only.” Responses were on 7-point scales from “strongly agree” to “strongly disagree” with neutral midpoints. Responses were rescaled ±3 with higher values indicating greater transphobia, then combined into an additive transphobia scale (α = 0.91, range: ±15, M = −0.55, SD = 8.65). In interactions, this 15-point additive scale was collapsed into a 7-point scale (range: ±3, M = −0.10, SD = 2.02) to give more observations per cell, with 0 indicating neutrality and the positive and negative values divided into thirds based on the scale range (±1/5, ±6/10, ±11/15). The full and the truncated additive scales yield substantively similar results.

Transgender Rights Policy Preferences

Respondents assessed nine policy statements: (a) “Legal protections that apply to gay and lesbian people should also apply to transgender people”; (b) “Congress should pass laws to protect transgender people from job discrimination”; (c) “Congress should not pass laws to protect transgender people from discrimination in public accommodations”; (d) “Insurance companies should not be required to pay for medical treatment related to transgender health issues”; (e) “Laws should protect transgender children from bullying in schools”; (f) “Businesses should have the right to refuse services to transgender people based on religious beliefs”; (g) “Transgender people deserve the same rights and protections as other Americans”; (h) “Transgender people should only be allowed to use public restrooms that are consistent with the sex listed on their driver’s license/state ID card”; and (i) “Transgender people should be allowed to serve openly in the military.” Responses were on 5-point scales from “completely agree” to “completely disagree” with a neutral midpoint. Items were rescaled ±2, coding +2 as the most pro-transgender position, and combined into an additive policy preference scale (α = 0.84, range: ±18, M = 3.98, SD = 7.43).

Feeling Thermometer

The survey included an ANES-style feeling thermometer battery asking respondents to rate groups on a 0 to 100 scale, with higher values indicating more positive feelings. From the groups assessed, our analysis used “music celebrities” (M = 53.58, SD = 25.49) as a control variable plausibly representing general interest in celebrities.

Knowing Transgender People

Respondents were asked, “Do you have a close friend or family member who is transgender?” (0 = no, 1 = yes; M = 0.07, SD = 0.25).

Age

Respondent age in years (M = 50.65, SD = 15.70) was collapsed into cohorts, unless otherwise noted, to increase observations per cell in interactions. Following a standard grouping in surveys, cohorts included 18 to 29, 30 to 44, 45 to 64, and 65+ years (scaled 1–4, M = 2.94, SD = 0.96). In the analyses, we compare the youngest and oldest cohorts to simplify discussion, though results for the middle two cohorts in all analyses fall between the reported values for the oldest and youngest.

Standard Covariates

Measured dichotomously were the following: sex (0 = male, 1 = female; M = 0.49, SD = 0.50), race (0 = White, 1 = non-White; M = 0.20, SD = 0.39), and LGBT identity (0 = non-LGBT, 1 = LGBT; M = 0.08, SD = 0.27). Household income ranged from “less than $19,999” to “$180,000 or higher” (scaled 1–10, M = 3.41, SD = 2.09). Education ranged from “less than high school” to “graduate degree” (scaled 1–5, M = 3.29, SD = 0.96). Church attendance ranged from “never” to “several times a week” (scaled 1–9; M = 3.93, SD = 2.88). Partisanship and ideology were measured with 7-point ANES scales; higher values, respectively, represented closer affiliation to the Republican Party (M = 3.72, SD = 2.02) and greater conservatism (M = 4.15, SD = 1.64).

Results

Following the Jenner Story

Our attention measure asked how much respondents “followed” the Jenner story. Like any news event, we cannot assume universal engagement with the story, so knowing which respondents were more likely to follow it may help explain evaluations of Jenner or associations between those evaluations and policy preferences. Nearly 80% of respondents followed the Jenner story to some degree: 22.69% followed it “not at all,” 31.99% “not too closely,” 31.08% “fairly closely,” and 14.24% “very closely.”

Table 1 models Jenner story attention. Beyond our hypotheses, associated with following Jenner more closely were the following: more frequent church attendance, higher income, knowing a transgender person, Democratic identification, warmer affect for music celebrities, and being female or LGBT.

Table 1.

Following the Jenner Story, 2015 CVR Survey.

	Following story
Transphobia	−.014* (.006)
Age	.001 (.002)
Church attendance	.076*** (.016)
Education	.067 (.046)
Female	.275** (.085)
Income	.087*** (.021)
LGBT	.412* (.167)
Non-White	.187^† (.113)
Partisanship	−.057* (.027)
Ideology	−.041 (.037)
Know a transgender person	.464** (.176)
Music celebrity thermometer	.013*** (.001)
τ₁	−.083
τ₂	1.458
τ₃	3.157
McFadden’s pseudo R²	.038
LR	195.910***
N	1,940

Note. Entries are weighted unstandardized ordinal logistic regression coefficients; standard error in parentheses. LGBT = lesbian, gay, bisexual, and transgender; LR = likelihood ratio.

†

p < .1. *p < .05. **p < .01. ***p < .001.

As noted, studies show that younger people tend to show greater interest in and attention to celebrities. Although H1 is the logical extension of that, the data did not support it. Age in years did not significantly predict following the Jenner story. Comparing average attention using the 4-point age cohort measure also showed no significant attention differences between cohorts, (F[3, 1937] = 1.12, p = .339).

As transphobia is a novel predictor in political behavior research, we first explore its nature. Figure 1 shows responses to the transphobia items, recoded into pro-transgender and anti-transgender attitudes. Two items elicited majority agreement from respondents. In the pro-transgender direction, 52.08% disagreed that “a man who dresses as a woman is a pervert.” In the anti-transgender direction, 55.08% agreed that “God made two sexes and two sexes only.” This question may indicate discomfort with gender fluidity or nonbinary gender identities, and the belief that gender and sex are indistinguishable.

Figure 1.

Response distributions of transphobia scale questions, 2015 CVR survey.

Respondents leaned in the pro-transgender direction on the other questions: 42.32% disagreed that “it is morally wrong for a woman to present herself as a man in public,” 44.32% would support a friend having gender reassignment surgery from male to female, and 41.07% disagreed that “sex change operations are morally wrong.” On the additive transphobia scale, its −0.55 mean indicated that sample on average held marginally non-transphobic attitudes. Overall, 46.87% of respondents earned negative scale scores indicating non-transphobic summary attitudes and 45.43% placed on the transphobic positive side of the scale. See Supplemental Appendix B for a separate analysis examining the correlates of the transphobia scale.

As we expect relatively transphobic individuals to be disinclined to engage with news about a transgender celebrity, H2 posited that transphobia would correlate negatively with following the Jenner story. Respondents at the transphobia minimum—the least transphobic—had a predicted probability of 17.54% (95% confidence interval (CI) [14.54, 20.55]) of not following the story at all, and a 15.54% (95% CI [12.72, 18.35]) chance of following it very closely. Among the most transphobic respondents, those respective probabilities were 24.52% (95% CI [20.59, 28.55]) and 10.75% (95% CI [8.51, 12.91]). Although H2 was supported, these predicted probability shifts in attention associated with transphobia were substantively modest. Furthermore, interactions with transphobia and other predictors did not yield moderating effects on attention.

Evaluation of the Jenner Story

We next examine how respondents fit the Jenner story into the broader social context. They could evaluate it as a change for the worse, an isolated incident, a change for the better, or indicate “no opinion.” The slight modal response—27.46% (N = 533)—was “no opinion,” though just 28.52% of these respondents did not follow the story at all. Thus, most lacking an opinion paid some attention to Jenner, but failed to contextualize what her story suggests for society for reasons indeterminable in our data. In surveys, Americans often lack opinions about high profile political matters, so the presence of these responses is unsurprising (Kleinberg & Fordham, 2017).

Among respondents who did not follow the Jenner story at all (N = 430), 64.65% still expressed an opinion about its social significance. Critically, our measure of following the story was not per se a measure of awareness about it. Some respondents could have known about the story from headlines or passive exposure, interpreted that as “not at all” following it if they did not actively seek that news, yet still have formed an opinion about it from this minimal awareness. To such respondents, Jenner may simply have been another transgender person coming out, with their evaluation of her merely reflecting gut perceptions of the significance of transgender people.

Table 2 models Jenner story evaluations among respondents who expressed opinions about it, dropping no opinion respondents who do not fit logically onto the dependent variable as scaled. Remaining responses were arrayed as an ordinal variable scaled ±1 and ordered from lowest to highest value as follows: change for the worse, isolated incident, and change for the better. The first model in Table 2 predicts evaluations as a function of demographic and political predictors, transphobia, and the measure of following the Jenner story. The music celebrity thermometer was dropped in this analysis given its less plausible connection to evaluating Jenner than paying attention to her.

Table 2.

Evaluation of Jenner Story, 2015 CVR Survey.

	Story Evaluation
Transphobia	−.691*** (.038)	−.275 (.226)
Following story	1.332*** (.156)	−.884^† (.535)
Age	−.216** (.066)	−.440* (.177)
Church attendance	−.039^† (.022)	−.034 (.022)
Education	.093 (.064)	.106 (.065)
Female	−.023 (.120)	.009 (.121)
Income	.028 (.029)	.037 (.029)
LGBT	.494* (.250)	.517* (.254)
Non-White	.194 (.165)	.204 (.168)
Partisanship	−.102** (.038)	−.106** (.038)
Ideology	−.104* (.052)	−.112* (.053)
Know a transgender person	.258 (.255)	.248 (.259)
Age × Follow	—	.793*** (.190)
Age × Transphobia	—	−.195* (.086)
Follow × Transphobia	—	−.822** (.250)
Age × Follow × Transphobia	—	.344*** (.093)
τ₁	.055	−1.687
τ₂	2.127	.411
McFadden’s pseudo R²	.291	.303
LR	897.140***	934.620***
N	1,408	1,408

Note. Entries are weighted unstandardized ordinal logistic regression coefficients; age is the collapsed 4-point cohort measure; transphobia is the collapsed 7-point scale; standard error in parentheses. LGBT = lesbian, gay, bisexual, and transgender; LR = likelihood ratio.

†

p < .1. *p < .05. **p < .01. ***p < .001.

H3 received support as age was negatively related to positive Jenner evaluations. All else equal, the predicted probability of saying that Jenner signaled change for the worse among 18- to 29-year-olds was 18.43% (95% CI [14.16, 22.71]), whereas the probability of calling her change for the better in this cohort was 35.79% (95% CI [29.50, 42.09]). Among those 65 plus, those probabilities were, respectively, 22.53% (95% CI [18.81, 26.25]) and 30.23% (95% CI [25.81, 34.64]). At both age extremes, then, labeling Jenner an isolated incident was the norm.

H4 also received support with more transphobic respondents less likely to evaluate Jenner positively. Transphobia more strongly differentiated Jenner evaluations than did age. The least transphobic respondents had a 72.48% (95% CI [67.82, 77.14]) predicted probability of saying that Jenner represented change for the better, and a 22.95% (95% CI [19.18, 26.72]) chance of calling her story an isolated incident. Conversely, the most transphobic had a 75.20% (95% CI [70.62, 79.78]) probability of saying that Jenner represented change for the worse, and a 20.81% (95% CI [17.10, 24.53]) chance of labeling her an isolated incident.

Further exploring key predictors, in examining whether a conditional relationship existed between age and transphobia, we also accounted for following the Jenner story given how many respondents did not follow it yet provided opinions about Jenner. The second model in Table 2 fits a triple interaction of age, transphobia, and following the story. Given our concern about “no attention” respondents, the attention variable for this interaction was dichotomized as following Jenner “not at all” (coded 0) versus following to any degree (coded 1; M = 0.77, SD = 0.41). Figure 2 plots this interaction for the youngest and oldest age cohorts.

Figure 2.

Evaluation of Jenner Story, Age × Follow Story × Transphobia Interaction, 2015 CVR survey.

Panel A in Figure 2 shows the effect of transphobia on Jenner evaluations among 18- to 29-year-olds who did not follow her story at all. Respondents in this grouping who also scored lowest in transphobia had a 49.01% (95% CI [24.59, 72.30]) chance of saying that Jenner represented change for the better and a 39.68% (95% CI [24.89, 5.47]) chance of calling her revelation an isolated incident. Among respondents in this group who scored highest in transphobia, their chance of labeling Jenner as change for the worse surged to 68.34% (95% CI [42.80, 93.87]).

Panel B in Figure 2 plots the transphobia effect among 18- to 29-year-olds who followed the Jenner story. Compared with Panel A, the key change here was for those lowest in transphobia. For them, attention to Jenner corresponded to a 78.61% (95% CI [70.33, 86.88]) chance of saying that Jenner represented change for the better—outpacing the 49% chance of their no attention counterparts. Among highly transphobic 18- to 29-year-olds, the probabilities of their various response options differed from those of their Panel A counterparts by about 1%—essentially, stability. Thus, among younger respondents, following Jenner only mattered for low transphobia individuals, raising the chance of them contextualizing Jenner in a more pro-transgender manner.

Panel C in Figure 2 graphs the transphobia effect among those 65+ who paid no attention to Jenner. Their pattern generally resembled that of inattentive 18- to 29-year-olds, except that these inattentive seniors were somewhat more likely to fit Jenner into normative perceptions of social change, evaluating her consistent with their transphobia scores. Low transphobia seniors had a 59.81% (95% CI [37.89, 81.73]) chance of saying change for the better, and high transphobia seniors had a 97.92% (95% CI [95.41, 100.00]) chance of saying change for the worse.

The more unique pattern in Figure 2 was Panel D, the transphobia effect among those 65+ who paid attention to Jenner. Low transphobia persons in this group had a 73.46% (95% CI [65.89, 81.03]) chance of calling Jenner change for the better—slightly less than 18- to 29-year-olds. However, high transphobia seniors were about as likely to label Jenner as change for the worse (46.37%; 95% CI [36.70, 56.03]) as they were an isolated case (41.64%; 95% CI [35.54, 47.74]). Indeed, among the highly transphobic in Figure 2, seniors who followed Jenner were least likely to see her as representing change for the worse and most likely to see her as an isolated case. In other groups, high transphobia translated to starker negative views of Jenner. Notably, highly transphobic and attentive seniors had just an 11.99% (95% CI [7.61, 16.37]) chance of calling Jenner change for the better, so any amelioration in their views of her associated with following her story was expressed as individuating her from social trends, and not explicit positivity.

Transgender Rights Policy Attitudes

As noted, the literature on parasocial interactions with LGBT figures disagrees over its effects on transgender rights attitudes. Examining whether evaluations of Jenner were associated with those policy preferences adds to that discussion. Our analysis again excluded those without substantive opinions about Jenner. That said, on our transgender rights policy scale, “no opinion” respondents on the Jenner evaluation item were on average more favorable toward transgender rights policies (M = 5.19) than those who expressed opinions (M = 3.59; t[1,939] = 4.24, p = .000). However, modeling policy attitudes as a function of our standard covariates and transphobia—minus Jenner-related variables—shows minimal differences in coefficient magnitude, statistical significance, and model fit between the full sample and the set of respondents expressing opinions about Jenner. This suggests that dropping no opinion respondents did not bias our analysis.

Table 3 models transgender rights policy attitudes. The first results column features no interactions. Predictors included demographics, political dispositions, transphobia, and evaluations of the Jenner story. Consistent with existing research, older respondents were less supportive of transgender rights. Those 65+ scored 2.2 units lower on the policy scale than 18- to 29-year-olds, or predicted values of 2.70 (95% CI [2.25, 3.15]) and 4.92 (95% CI [3.31, 5.53]), respectively, other predictors at their means.

Table 3.

Transgender Rights Policy Preferences, 2015 CVR Survey.

	Policy Scale
Transphobia	−1.806*** (.096)	−2.149*** (.275)
Story evaluation	1.813*** (.224)	.469 (.683)
Age	−.738*** (.154)	−.510* (.198)
Church attendance	−.062 (.053)	−.048 (.052)
Education	−.365* (.152)	−.336* (.151)
Female	.671* (.283)	.775** (.283)
Income	.102 (.068)	.105 (.068)
LGBT	.359 (.529)	.576 (.528)
Non-White	.081 (.383)	−.052 (.383)
Partisanship	−.449*** (.091)	−.454*** (.090)
Ideology	−.287* (.122)	−.311* (.122)
Know a transgender person	.449 (.554)	.574 (.550)
Age × Evaluation	—	.474* (.238)
Age × Transphobia	—	.126 (.094)
Evaluation × Transphobia	—	−.128 (.288)
Age × Evaluation × Transphobia	—	.206* (.100)
Constant	9.382*** (.829)	9.052*** (.887)
Adjusted-R²	.584	.592
F	165.700***	128.330***
N	1,408	1,408

Note. Entries are weighted unstandardized ordinary least squares regression coefficients; age is the collapsed 4-point cohort measure; transphobia is the collapsed 7-point scale; standard error in parentheses. LGBT = lesbian, gay, bisexual, and transgender.

†

p < .1. *p < .05. **p < .01. ***p < .001.

The data supported H5 in that more transphobic respondents were less supportive of transgender rights. On the collapsed 7-point transphobia scale, the most transphobic respondents placed 10.83 units lower—less supportive of transgender rights—on the policy scale than the least transphobic respondents, with predicted values of −1.85 (95% CI [−2.48, −1.22]) and 8.98 (95% CI [8.36, 9.60]), respectively, holding other predictors at their means.

H6 also received support as more positive evaluations of Jenner were associated with stronger support for transgender rights policies. Respondents who said that the Jenner story was indicative of change for the better were predicted to score 5.47 (95% CI [4.94, 5.99]) on the policy scale, holding other predictors at their means, whereas those who labeled it change for the worse had a predicted policy score of 1.84 (95% CI [1.34, 2.34]).

Exploring conditional effects, since age and transphobia had a moderated relationship with following the Jenner story in predicting story evaluations, the second model in Table 3 fits a triple interaction of age and transphobia with Jenner story evaluations to test for a moderated relationship in predicting policy support. This yielded a significant moderated effect. For 18- to 29-year-olds, the three lines in Panel A of Figure 3 represent predicted transgender rights policy support (Y-axis) for the three Jenner story evaluations with the transphobia minimum and maximum values on the X-axis. Predicted policy support for the least transphobic 18- to 29-year-olds ranged from 10.44 to 11.86, and declined by about 12 points in each Jenner story evaluation category when compared with the transphobia maximum value. For the youngest respondents, then, Jenner evaluations were irrelevant to policy support, with transphobia exhibiting a clear direct and negative effect.

Figure 3.

Transgender rights policy preferences, Age × Story Evaluation × Transphobia Interaction, 2015 CVR survey.

That was not the case for those age 65+ in Panel B. Like their younger counterparts, Jenner evaluations were irrelevant to transgender rights support among the least transphobic seniors where predicted support ranged from 8.20 to 8.75 across the three Jenner evaluations. However, as seniors scored higher on transphobia, evaluations of Jenner became a stronger moderator of how much transphobia depressed policy support. For seniors who thought that Jenner represented change for the worse, their predicted transgender rights support declined 14.03 units comparing the least and the most transphobic. That same decline was 9.86 units for those who saw Jenner as an isolated case, and 5.67 units for those who said that she represented change for the better.

As discussed, most highly transphobic seniors said that Jenner was either change for the worse or an isolated case. Predicted policy support for these two types of respondents, respectively, was −5.83 (95% CI [−6.86, −4.80]) and −1.38 (95% CI [−2.45, −0.29]). Therefore, while seeing Jenner as an isolated case did not on balance translate to pro-transgender policy preferences among the most transphobic seniors, the 4.45 unit difference in these two predicted scores did represent a marked substantive difference give the range of the policy scale.

Notably, younger respondents at the transphobia minimum always exhibited higher predicted policy support than their 65+ counterparts no matter how they evaluated Jenner, with this difference approximately 2 units. Among the most transphobic respondents, 18- to 29-year-olds who saw Jenner as change for the worse scored about 2.5 units higher on the transgender rights scale than those 65+ who thought the same. But for the minority of highly transphobic respondents who saw Jenner as change for the better, it was those 65+ who had higher predicted support for transgender rights than those in the youngest cohort—a nearly 3 unit difference.

Also, the conditional effects of transphobia in analyses here were unique to that scale. In models not shown, we replicated our analyses, replacing transphobia with a transgender feeling thermometer and an ordinal comfort with transgender people measure. Neither replicated the transphobia results. This suggests that, rather than being a measure of mere pro- or anti-transgender sentiment, the transphobia scale captured some unique element of how respondents perceived transgender persons, and how those perceptions interacted with the Jenner story.

Discussion

When Caitlyn Jenner came out as transgender in 2015, Americans refamiliarized themselves with her at a time when transgender rights were becoming more politically salient. Our data suggest that Jenner’s identity revelation was politicized for many Americans. Attention to her story and evaluations of its significance were divided by partisanship, ideology and transphobia. Furthermore, transphobia combined with age and attention to the story to shape how individuals perceived it in the context of broader social change around transgender persons. Those perceptions of her story were related to support for transgender rights, additionally interacting with age and transphobia to affect policy preferences.

Our findings have several important implications. First, our results remind political science that celebrities merit scholarly attention. Given fascination with celebrity culture, many Americans may be more familiar with and pay more attention to celebrities than politicians with formal political power. Many celebrities may have significant power to capture citizen attention with a political statement, or to grab the attention of citizens who are not routinely attentive to politics.

Yet, celebrities have received scant attention in political behavior research. That omission is understandable given the field’s reliance on large and expensive national surveys which may take months to develop and over which researchers may have no influence on the content. When a celebrity, sometimes unpredictably, becomes politically relevant, analysts may not have the time, money, or personal influence over other researchers to get “gold standard” data from a nationally representative sample. If political behavior scholars are to study celebrities seriously, then research may have to rely more on experiments, convenience samples, or nonsurvey-based methods that both dominate research on celebrity effects in other fields and make that research more practical.

Second, scholars should appreciate different ways that celebrities can be political. Research typically focuses on celebrity endorsements, and these overt attempts at mass persuasion certainly merit attention. However, that mold of celebrity activism did not fit Jenner, at least at first. Nor does it fit celebrities whose creative efforts like music, movies, and fashion, or their identity are often discussed through a political lens. Jenner’s political act in 2015 was to be her genuine self publicly, an act that may raise awareness of the small and marginalized transgender community, and that can be seen both a personal declaration of identity and as part of a broader political strategy to promote change. Especially if a celebrity becomes the public face of a politically relevant group like transgender people with whom most Americans may not have regular contact, these public declarations of identity and their resulting parasocial contact effects demand serious political inquiry, as our work demonstrates.

Third, the parasocial contact effect that we document underscores the critical role of mass media portrayals of transgender persons—real or fictional—in shaping transgender rights attitudes. These portrayals can be politically impactful and are likely to reach larger audiences than transgender Americans can easily reach in face-to-face social interactions given their small numbers. However, for that exposure to become explicit political advocacy, the burden is on celebrities to use their fame as a political platform, which may threaten their fame itself. And given the power than an individual transgender celebrity may have to “represent” transgender Americans to people who may not otherwise know a transgender person, those celebrities have substantial power to control the issue agenda when they do advocate. The choice of what they discuss publicly—their own identity and life, versus systemic issues like violence or discrimination—may be unusually important for shaping what others perceive as key issues facing transgender people.

The media is also complicit in amplifying certain celebrity voices. Media decide to ignore or publicize the advocacy of transgender celebrities, a choice that celebrities do not control. With a small group like transgender Americans, the choice of which celebrities to cover is important for shaping mass perceptions of the group and its politics. Someone like Jenner who carries racial and economic privilege and espouses conservative politics may convey a different perception about the diversity of transgender Americans and what issues affect them than another celebrity might.

Fourth, we have shown for the first time in political behavior literature that transphobia, per Hill and Willoughby (2005), influences transgender rights policy attitudes. For many social groups, scholars have developed measures that assess forms of bias particular to that group. For example, gender role traditionalism scales or symbolic racism tap unique ways that women and African Americans, respectively, are politicized. Similarly, the transphobia scale focuses on the unique matters of transgender identity that often become politically contentious, and the paucity of political research on it opens many avenues for future research.

Of course, there are limitations to our research. Although our survey provides strong external validity since respondents were answering questions about Jenner contemporaneous with her coming out, our analysis can only observe associations between variables because selection bias drives attention to her story. Although that is certainly how news attention works in the real world and there is value in surveys like ours that capture celebrity phenomena in real time, experiments can provide better evidence of causal mechanisms driving celebrity effects on political behavior.

Relatedly, celebrity studies from other fields often consider traits unique to particular celebrities—likability, credibility, or perceived similarity, for example—that can influence the persuasiveness of an endorsement. Our survey included no such measures about Jenner, so we cannot assess how qualities unique to Jenner affected the outcomes that we observed, if at all.

Finally, every celebrity has unique biographies, traits, and demographics that may cause their influence to vary. Should a study find that a celebrity seemingly influences mass behavior, it can be difficult to determine whether those effects are generalizable to other celebrities. We cannot say, had we studied another transgender celebrity, whether we would have observed the same relationships that we did. Nor can we say whether other transgender celebrities would influence the same types of people. If Jenner had an “audience” that seemed more responsive to her story, it was older respondents—particularly senior citizens like Jenner herself who may have been familiar with her for decades—who were highly transphobic. Other celebrities may have different audiences who would be more responsive to their unique appeal. In that sense, every celebrity may be a unique case study, which deters taking celebrity effects seriously in a field like political science where systematic and generalizable findings are often preferred over findings that are unique to particular high-profile figures, at least outside of presidents.

Furthermore, barring replication, Jenner may not even affect the same audience post-2015 given how her narrative has evolved. Her high-profile entanglement in the 2016 presidential election, particularly with Donald Trump and conservative politics, occurred after our survey. Those events may have explicitly politicized Jenner along partisan or ideological lines in a way that she was not politicized simply by coming out in 2015. If her actions surrounding the election altered how she was politically perceived by certain types of Americans, then any parasocial effects on transgender rights attitudes tied to her transgender identity may manifest differently today, if at all.

Despite these limitations that necessitate further research, our findings should encourage political scientists—and not just those working on LGBT issues—to take celebrities seriously as forces who can exert real influence on average citizens. Ours is a study of one celebrity in one context defined by biography and time, but that is fundamentally what celebrity politics is: one person being political in a specific circumstance. Only by seriously treating celebrities as political actors and studying them through multiple methods and replication can we understand how much of the power of celebrity stems from fame itself versus individual personality.

Supplemental Material

supplemental_material – Supplemental material for The Politics of Being “Cait”: Caitlyn Jenner, Transphobia, and Parasocial Contact Effects on Transgender-Related Political Attitudes

Supplemental material, supplemental_material for The Politics of Being “Cait”: Caitlyn Jenner, Transphobia, and Parasocial Contact Effects on Transgender-Related Political Attitudes by Patrick R. Miller, Andrew R. Flores, Donald P. Haider-Markel, Daniel C. Lewis, Barry Tadlock and Jami K. Taylor in American Politics Research

Footnotes

Declaration of Conflicting Interests

The author(s) declared no potential conflicts of interest with respect to the research, authorship, and/or publication of this article.

Funding

The author(s) received no financial support for the research, authorship, and/or publication of this article.

ORCID iDs

Patrick R. Miller

Barry Tadlock

Supplemental Material

Supplemental material for this article is available online.

Author Biographies

Patrick R. Miller is an associate professor in the Department of Political Science at the University of Kansas. His research focuses on American political behavior and LGBT politics, primarily from a psychological perspective.

Andrew R. Flores is an assistant professor at American University and a Visiting Scholar at the Williams Institute, UCLA School of Law. He is a member of the National Academies of Sciences, Engineering, and Medicine Consensus Committee on the Status and Well-being of Sexual and Gender Diverse Populations in the United States.

Donald P. Haider-Markel is professor and chair of political science at the University of Kansas. His research and teaching is focused on the representation of group interests in politics and policy.

Daniel C. Lewis is an associate professor and chair of the Department of Political Science & International Relations at Siena College. His research examines representation in U.S. politics, with a particular emphasis on policy responsiveness and minority rights.

Barry Tadlock is professor of Political Science at Ohio University. His research interests include LGBTQ-related public opinion and public policy.

Jami K. Taylor is a professor of political science and public administration at the University of Toledo. Her research interests are in the area of LGBTQ politics, and policy diffusion.

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