架け橋としての連帯：共有された差別は有色人種の投票意図と間接的に関連する

Introduction

The year 2020 marked a period of profound upheaval in U.S. society for people of color (PoC). As the COVID-19 pandemic spread, rhetoric blaming China for the virus became widespread in political discourse (Martin & Haberman, 2020; Neuman, 2020), contributing to a surge in anti-Asian prejudice and hate crimes (Lantz et al., 2022; Lim et al., 2022; Reny & Barreto, 2020; Ruiz et al., 2023). Meanwhile, the police killings of Breonna Taylor and George Floyd laid bare the systemic injustices that Black Americans continue enduring within U.S. society and its criminal justice system (Toosi et al., 2021; Toraif et al., 2023). These events unfolded amid a broader climate of heightened immigration restrictions and policies that disproportionately affected Muslim and undocumented Latino communities (Hamedy, 2018; Kang & Yang, 2021), during an election cycle that amplified debates about race, identity, and national belonging (Sides et al., 2022). Together, these developments underscored that racial minorities remain embedded within a durable racial hierarchy—continually relegated to second-class citizenship—while also highlighting the need for PoC to unite politically in their fight for equality and fair treatment (Pérez, 2021; Starr, 2023).

Drawing on social identity theory, researchers have sought to foster intraminority solidarity by making experiences of shared discrimination salient (Chaney et al., 2018; Cortland et al., 2017; Craig & Richeson, 2012, 2016; Craig et al., 2012; Pérez, Vicuña, & Ramos, 2024). This approach is based on the premise that highlighting discrimination against an ingroup can heighten perceived similarity with outgroups that face comparable marginalization, thereby promoting unity under a shared, superordinate identity. For instance, Schmitt et al. (2003) found that international students from 32 countries were more likely to identify collectively as “international students” when they perceived nationality-based discrimination. Yet, making discrimination salient does not always yield solidarity: when groups are marginalized along different dimensions (e.g., race vs. gender), it can instead evoke perceived threat and outgroup derogation (Craig & Richeson, 2012, 2016; Craig et al., 2012). Thus, discrimination salience promotes solidarity only when groups perceive themselves as similarly stereotyped or positioned within a social hierarchy (Chaney et al., 2018).

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To mitigate this risk, explicitly linking the experiences of disadvantaged ingroups and outgroups can transform threat into coalition (Craig & Richeson, 2016). For example, Black participants who read about the historical parallels between racial and sexual discrimination expressed greater empathy for sexual minorities and stronger support for same-sex marriage (Cortland et al., 2017). Building on this idea, subsequent research has used Zou and Cheryan’s (2017) Racial Position Model to specify the axes of discrimination that are most likely to unite groups. This work finds that cross-group solidarity is most likely when groups share a common axis: Asians and Latinos—often racialized as foreign or un-American—express greater solidarity when discrimination based on perceived foreignness is emphasized, whereas Latinos and Black Americans— often stereotyped as socially inferior—show stronger solidarity when subordination is highlighted (Chin et al., 2022; Cortland et al., 2017; Eidgahy & Pérez, 2022; Pérez, Vicuña, & Ramos, 2024).

Despite growing evidence for shared discrimination appeals, little is known about whether and when such appeals mobilize solidarity between Asian and Black Americans, two communities racialized along distinct historical and social axes (Nicholson et al., 2025). Asian Americans are often stereotyped as socially superior to other PoC (Zou & Cheryan, 2017) and framed as “model minorities,” in contrast to Black Americans, who are more frequently stereotyped as socially inferior (Lee et al., 2009). This “competitive victimhood” narrative casts racial progress as a zero-sum endeavor and undermines the development of mutual recognition and perceived linked fate—psychological foundations of sustained coalition-building (Goh & Douglas, 2025; McNeill & Vollhardt, 2020; Nicholson & Mei, 2023; Noor et al., 2012; Rogbeer et al., 2025). Historically, such narratives have not only impeded solidarity but have also fueled intergroup tension. For example, long-standing economic and social frictions between Black and Korean American communities in Los Angeles, Chicago, and New York culminated in violent conflict during the early 1990s (Kim, 2000). More recently, viral reports of assaults on Asian elders and women—often attributed to Black perpetrators—have reignited perceptions of mutual hostility and reinforced the belief among some Asian Americans that Black communities harbor anti-Asian prejudice (Lee & Huang, 2021).

Research on political solidarity—defined as a sense of shared commitment and collective responsibility among members of different marginalized groups to advance one another’s social and political standing—has largely focused on intergroup attitudes and policy support rather than on political behavior. Yet lasting structural change requires more than shifting opinions; it depends, critically, on the capacity to translate attitudes into collective political action. Theories of reasoned action and planned behavior identify behavioral intentions as one of the strongest predictors of behavior, a finding supported across a range of behavioral domains (e.g., blood donation, energy saving, contraceptive use) with average intention–behavior correlations around r ≈ .50 (Ajzen, 1985; Armitage & Conner, 2001; Kim & Hunter, 1993; Randall & Wolff, 1994; Webb & Sheeran, 2006). Even in voting research, intentions are among the strongest correlates of turnout (Glaser, 1958; Kim & Hunter, 1993; Ko et al., 2025; Randall & Wolff, 1994). Although pre-election intentions often overestimate actual participation (Achen & Blais, 2016), they remain a meaningful indicator of political engagement among PoC (e.g., Pérez, 2015; Pérez & Mártir Luna, 2025).

The present study examines whether making shared discrimination salient can catalyze voting intentions among nationally representative samples of Black, Latino, and Asian Americans in the lead-up to the 2024 U.S. presidential election. Specifically, we test whether heightened feelings of solidarity predict intentions to vote for (i) a PoC representative, defined as a presidential candidate perceived to advocate for PoC interests, and (ii) Democratic candidate Kamala Harris— the first woman of color to secure a major-party presidential nomination while representing a party broadly viewed as the “home” of racially minoritized communities (Mason, 2023). Leveraging three parallel survey experiments with Black, Asian, and Latino American adults, we compare the strength and direction of solidarity’s influence on voting intentions across racial groups and assess the robustness of these relationships through sensitivity analyses. Below, we draw on social identity theory to explain how shared discrimination may promote solidarity on behalf of PoC—a meaningful superordinate category with political implications (Pérez, 2021; Pérez et al., 2025)—and specify when effects may diverge, reflecting the distinct histories and perceived prototypicality of Asian, Black, and Latino Americans within the broader PoC category.

Activating Solidarity: Promise and Precarity in Cross-Racial Appeals

Research on racial coalitions in the U.S. builds on the well-established principle that perceived similarity underlies positive intergroup attitudes and prosocial outcomes (Eisenberg & Miller, 1987; Krebs, 1975; Lemay & Ryan, 2020; Miller et al., 2011; Schütt, 2023). Consistent with this perspective, Goldstein et al. (2014) demonstrated that believing someone has taken one’s perspective increases perceived similarity. This sense of shared experience, in turn, fosters liking and prosocial behavior toward that person. The benefits of these interpersonal processes can generalize to the broader groups to which the individuals belong, reducing prejudice and stereotyping toward the group as a whole (e.g., Batson et al., 1997; Broockman & Kalla, 2016; Galinsky & Ku, 2004; Galinsky & Moskowitz, 2000; Vescio, Sechrist, & Paolucci, 2003). Collectively, this work suggests that perceiving others as targets of similar discrimination may provide a powerful psychological basis for empathy and coalition formation among racial groups.

Building on this foundation, researchers have explored how making discrimination salient shapes intergroup attitudes. Two contrasting perspectives derived from social identity theory (Tajfel & Turner, 1979) offer competing predictions about these effects and have been tested across multiple experiments (e.g., Craig & Richeson, 2012, 2014, 2016; Craig et al., 2012). The threat hypothesis posits that highlighting discrimination can threaten the ingroup’s social standing, prompting outgroup derogation as a means of restoring group esteem (Branscombe et al., 1999; Craig & Richeson, 2014; Craig et al., 2012). For example, when the social and economic consequences of racial discrimination were made salient, heterosexual Black and Latino participants reported more negative attitudes toward sexual minorities—a pattern also observed in two nationally representative samples of Asian Americans (Craig & Richeson, 2014). Similarly, making sexism salient for White women increased racial bias towards Black and Latino individuals (Craig et al., 2012).

In contrast, the solidarity hypothesis suggests that making discrimination salient can enhance perceptions of similarity with other marginalized groups, leading to positive intergroup outcomes. For example, White American women experienced less identity threat when interacting with a Black male expert (compared to a White male expert) because they perceived him as sharing a disadvantaged social position (Chaney et al., 2018). Extending this logic, several experiments demonstrate that perceiving discrimination experiences as shared consistently improves intergroup perceptions (Cortland et al., 2017; Craig & Richeson, 2012, 2016; Jun et al., 2023). In one study, making anti-Asian racism salient led Asian Americans to express more positive explicit and implicit attitudes toward Black Americans—an effect partially mediated by perceived similarity (Craig & Richeson, 2012). Similarly, Galanis and Jones (1986) found that Black participants who read an insanity plea linking racism to mental illness subsequently endorsed fewer negative stereotypes of a person with mental illness, illustrating how recognizing shared oppression can reduce stigma across identity domains.

Although racial minorities might be assumed to share a common identity dimension, solidarity is more likely to emerge when distinct subgroups (e.g., Black and Latino people) are explicitly encouraged to view themselves as part of a broader superordinate identity such as people of color (Craig & Richeson, 2016; Pérez, 2021). This process aligns with the Common Ingroup Identity Model (Gaertner et al., 1989; Gaertner & Dovidio, 2000), which posits that highlighting shared similarities encourages individuals to recategorize themselves as members of a superordinate group. Such recategorization reframes collective challenges from a “them” issue to an “us” issue, fostering empathy, cooperation, and intergroup solidarity (Ball & Branscombe, 2019). Moreover, when discrimination is understood as a shared grievance rooted in common sources of oppression, it can also evoke anger and resentment—emotions that drive mobilization under a politicized collective identity (Lalonde & Cameron, 1994; Simon & Klandermans, 2001; van Zomeren et al., 2008; Warner et al., 2014). These politicized identities, in turn, shape justice concerns (Tyler & Smith, 1999) and motivate political engagement aimed at challenging systemic inequities (Simon & Klandermans, 2001).

Recent experimental work extends this framework from attitudes to political preferences. Black, Asian, Latino, and MENA Americans were randomly exposed to news articles highlighting shared experiences of discrimination with a target outgroup, framed as stemming from perceived inferiority or foreignness– in line with Zou and Cheryan’s (2017) framework (Chin et al., 2022; Cortland et al., 2017; Craig & Richeson, 2016; Eidgahy & Pérez, 2022; Pérez, Vicuña, & Ramos, 2024). Participants who read these shared discrimination narratives reported greater solidarity with PoC and stronger support for policies benefiting other racial minorities. For instance, Black participants who read about shared second-class citizenship with Latinos subsequently expressed greater support for policies benefiting Latinos (e.g., “Introducing harsher penalties for hate crimes against Latinos”). Meta-analytic evidence (Pérez, Vicuña, & Ramos, 2024) indicates that shared discrimination, on average, increases solidarity between people of color (d = 0.18) and that solidarity, in turn, predicts support for outgroup-advantaging policies (d = 0.79). Sensitivity analyses suggest this downstream association is robust to unmeasured confounding (ρ = .330), such that solidarity would have to be moderately correlated to a confounder for its effect on policy support to be reduced to zero. Using a design-based blockage manipulation intended to neutralize the influence of solidarity (MacKinnon & Pirlott, 2014), Rogbeer et al. (2025) found that solidarity’s effect on policy support, while modestly reduced, remained largely intact—bolstering confidence in its causal role.

Despite these encouraging findings, the unifying potential of shared discrimination is conditional. Social categorization is fluid and context-dependent, shaped by cues in the immediate social environment (Turner et al., 1987). In the U.S., race remains a dominant axis of social organization (Zou & Cheryan, 2017) and group-specific histories and lived experiences often become more psychologically salient than shared marginalization, reinforcing intergroup boundaries and limiting durable coalition even when shared oppression is acknowledged (e.g., Brilliant, 2010; Craig & Richeson, 2016; McClain et al., 2009; Meier et al., 2004; Vaca, 2004). These divisions are compounded by divergent historical legacies (e.g., the Chinese Exclusion Act vs. the one-drop rule) (Ngai, 2004), competition for resources and recognition (e.g., Students for Fair Admissions v. Harvard) (Barnes & Moses, 2021), and racialized narratives such as the “model minority” stereotype (Tuan, 1999), which can obscure systemic inequality and cast some groups as inherently distinct.

These dynamics help explain why appeals to a superordinate identity do not resonate uniformly. Asian Americans, for instance, are often triangulated between the White majority and other racial minority groups (Craig & Lee, 2022; Kim, 1999; Wang & Santos, 2023; Zou & Cheryan, 2017), complicating their alignment with a broader PoC identity. Like Latinos, Asians experience discrimination rooted in perceptions of foreignness; unlike Black and Latino Americans, they are simultaneously stereotyped as superior—a position they share with the White majority (Zou & Cheryan, 2017). This dual positioning can produce ambiguous political allegiances, making solidarity less predictable and more issue dependent (Craig & Lee, 2022).

The perception of Asian Americans as “superior” is largely driven by the model minority myth, which attributes their success to cultural values and a strong work ethic. Social identity theory holds that individuals are motivated to maintain a positive social identity (Tajfel, 1978; Turner & Brown, 1978), a motive that can legitimize status hierarchies. This suggests that some Asian Americans may endorse this seemingly positive stereotype, with downstream effects on intergroup attitudes. Indeed, Asian American participants who endorse the model minority myth were found to hold more anti-Black attitudes, oppose affirmative action, and perceive lower levels of discrimination (Le et al., 2024; Yi & Todd, 2021; Yi et al., 2023). Consistent with this logic, Pérez et al. (2025) show that conservative Asian Americans primed with the model minority narrative express less solidarity with other PoC, distancing themselves from a shared racial identity. These dynamics were evident in the Supreme Court’s recent decision on affirmative action in university admissions, framed by some as redressing anti-Asian discrimination. The lawsuit — spearheaded by a White conservative activist and presented by anonymous Asian plaintiffs — ultimately succeeded in dismantling race-conscious admissions, reinforcing meritocratic narratives and, for some, affirming beliefs in Asian exceptionalism even as it upheld structural anti-Blackness and white supremacy (Kim, 2023; Liu et al., 2023).

Taken together, these patterns clarify why recategorization under a superordinate identity is not equally effective across groups. Perceptions of prototypicality and belonging within superordinate categories like PoC can significantly moderate the impact of shared discrimination appeals. Superordinate labels may encourage temporary identification and a sense of linked fate in experimental contexts– especially when discrimination is framed as a shared grievance– but enduring coalitions require deeper identification with the superordinate group. According to self-categorization theory, individuals assess their fit with group norms and prototypes to determine belonging (Turner & Reynolds, 2012). In this context, the degree to which Black, Latino, and Asian Americans see themselves as prototypical PoC likely shapes how strongly they internalize shared discrimination messages. Pérez (2021) finds that Black Americans are widely perceived as the most prototypical members of the PoC category, followed by Latinos and then Asians: a consensus that emerges across members of these racially stigmatized groups. Consequently, Latinos and especially Asians may feel less psychologically included in the broader PoC identity, reducing the motivational impact of shared discrimination frames on intergroup solidarity (Chin et al., 2022).

Taking PoC Solidarity to the Ballot Box in an Age of Political Polarization

Variation in racial positions suggest that not all groups will be equally compelled by shared discrimination appeals—a distinction that may have meaningful implications for political behavior in a polarized electoral context (but see Engelhardt et al., 2025; Pérez, Goldman et al., 2024). Scholars generally agree that most voters are cognitive misers (Fiske & Taylor, 1991; Lau & Sears, 1986), meaning they rely on mental shortcuts rather than engage in extensive information processing when making political decisions. One such heuristic is party stereotypes, which help voters quickly assess candidates and policy positions (Hayes, 2011; Lau & Redlawsk, 2001; Philpot, 2004). As a result, party affiliation serves as a primary lens through which voters interpret political information, shaping perceptions of candidates and their stances on key issues (Conover & Feldman, 1989; Hamill et al., 1985). Given this reliance on cognitive shortcuts, it is essential to understand how political elites have influenced the development of party stereotypes—stereotypes that, in turn, shape voters’ perceptions of which party and candidate are most likely to represent the interests of PoC in national elections.

In 1964, President Johnson signed the Civil Rights Act into law, marking a pivotal moment that triggered a major realignment in U.S. politics (Black, 2004; Carmines & Stimson, 1989). In the ensuing years, the Democratic Party solidified its association with civil rights by championing policies that expanded fair housing protections, desegregated schools, and safeguarded Black voting rights, thereby establishing itself as the preferred party for racial minorities (Gilens et al., 1998; Weissberg, 1991). More recently, the Democratic Party has further solidified its association with marginalized groups by positioning itself as the party of Black, Latino, and Asian people, with self-reports and indirect measures (e.g., IATs) corroborating this mental link between various PoC and Democrats (Pérez et al., 2025; Westwood & Peterson 2022; Zhirkov & Valentino, 2022).

Conversely, the Republican Party has gradually positioned itself as the ideological opposite of Democrats on racial matters (Heit & Nicholson, 2010; Hout & Maggio, 2021; Levendusky, 2009; Mason, 2018). This shift became evident during Nixon’s 1968 campaign when the Republican National Committee deliberately removed any mention of civil rights from its platform (Pearcy & Clabough, 2019). Key strategist Kevin Phillips underscored this realignment in a New York Times interview, noting that Republicans were not interested in courting Black American voters, as “the more Negroes who register as Democrats in the South, the sooner the Negrophobe whites will quit the Democrats and become Republicans” (Boyd, 1970). This trend continued during Barack Obama’s presidency, with Republican elites further solidifying their anti-immigrant and racially exclusionary image by spreading false claims about Obama’s citizenship and religious affiliation (Hughey & Parks, 2014; Tesler, 2016; Tesler & Sears, 2010; Wise, 2021). By Donald Trump’s first presidency in 2016, the Republican Party had taken an even harder stance on immigration, with 82% of Republican House members voting to reduce legal immigration by 40%, marking the most significant proposed reduction since the national-origin quotas of the 1924 Immigration Act (Johnson, 2018).

This growing racial and ideological divide between the two parties has strengthened the association of the Republican Party with racial conservatism and the Democratic Party with racial liberalism (Mason, 2015, 2016). Thus, in America’s two-party system, it is generally the case that Black, Latino, and Asian voters perceive Democratic candidates as more likely to advocate for their racial ingroup’s interests (Abrajano & Alvarez, 2010; Kuo et al., 2017; White & Laird, 2020). Indeed, even in the 2024 election, despite some support for Republican Donald J. Trump among Asian American and Latino voters, the center of political gravity among African Americans, Asian Americans, and Latinos remains firmly pro-Democratic (Fraga et al., 2025; Pérez et al., 2025). Additionally, given that a significant proportion of Hispanics (33%) and Asians (65%) are foreign-born (Batalova, 2025; Ramakrishnan et al., 2025), and that the majority of undocumented immigrants come from Mexico (Passel & Krogstad, 2024), President Trump’s regular use of racist and xenophobic rhetoric has solidified the perception among these communities that the Republican Party does not represent their interests (Abramowitz & McCoy, 2019; Pérez et al., 2025; Philpot, 2004; White & Laird, 2020).

This history of party polarization leads us to anticipate that Black, Latino, and Asian adults who feel greater solidarity with PoC will be more likely to support Democratic candidates, whom they perceive as friendlier to PoC rights. However, because Black Americans already display high and lopsided baseline support for Democratic candidates (White & Laird, 2020), their vote intentions may be less sensitive to fluctuations in solidarity. In contrast, Asian and Latino voters—whose partisan affiliations are more variable—may be more responsive to solidarity cues when making electoral decisions (Abrajano & Alvarez, 2010; Wong et al., 2011). In short, racial group membership may moderate solidarity’s influence on voting intentions.

The Current Research

Although prior work demonstrates that shared discrimination heightens solidarity among PoC, which in turn increases support for policies benefiting other marginalized groups (Cortland et al., 2017; Craig & Richeson, 2016; Kim et al., 2025; Pérez, Vicuña, & Ramos, 2024), it remains unclear whether such appeals are potent enough to influence concrete political behavior in high-stakes, real-world elections. To address this gap, we conducted three large-scale experiments with nationally representative samples of Black (N = 850), Latino (N = 850), and Asian American (N = 850) adults three weeks before the 2024 U.S. presidential election between Democrat Kamala Harris and Republican Donald Trump. Leveraging this real-world electoral context allows us to assess whether exposure to shared discrimination is indirectly associated with voting intentions—an established behavioral proxy that remains moderately correlated with turnout over time (Kim & Hunter, 1993; Randall & Wolff, 1994)—through heightened solidarity with PoC.

Prior research indicates that voters often infer representativeness from shared social identities (Bejarano et al., 2021; Reher & Evans, 2025). For example, Black and Latino voters were more likely to perceive candidates who shared their racial or gender identities as advocates for their community’s interests, relative to candidates with whom they did not share an identity (Bejarano et al., 2021). The Democratic Party has long been regarded as the political “home” of racially minoritized communities (Mason, 2023) because of its historical support for policies promoting racial equality (Philpot, 2017; Philpot & Walton, 2007). This reputation may shape how voters of color interpret identity cues: when a Democratic candidate shares their racial or gender identity, that overlap may reinforce perceptions that the candidate will advocate for their group’s interests. Therefore, we measured voting intentions using two parallel outcomes. The first captured respondents’ intent to vote for a PoC representative—defined as a candidate perceived to advocate for the interests of people of color. The second assessed intent to vote for Kamala Harris, the first woman of color to secure a major-party presidential nomination. By including both outcomes, our design directly tests whether solidarity with people of color predicts support for candidates perceived as representing PoC interests—and whether Harris successfully positioned herself as such during the 2024 election.

Using our parallel experimental design, we test whether solidarity with PoC mediates the effect of shared discrimination on voting intentions, and whether this indirect effect varies by race. Drawing on research on group prototypicality and superordinate identity (Pérez, 2021), we expect that Black participants—often viewed as the most prototypical members of the PoC category—will show the strongest solidarity response to shared discrimination. Conversely, Asian Americans, whose racial position is shaped by the dual forces of the model minority stereotype and perceived foreignness, are expected to exhibit the weakest solidarity response. We further anticipate that racial group membership will moderate the relationship between solidarity and voting intentions. Given that Black Americans are often seen as the prototypical PoC and that voters frequently interpret shared group identities as signals of political representativeness (Bejarano et al., 2021; Reher & Evans, 2025), increases in solidarity should be more strongly associated with intent to vote for a PoC representative among Black Americans than among Latino and Asian Americans. However, because Latino and Asian Americans’ partisan preferences tend to be more fluid, we expect solidarity with PoC to be more strongly associated with intent to vote for Harris in these groups than among Black Americans, whose Democratic alignment is already well established.

Methods

Participants

Published research on the indirect effect of shared discrimination on policy preferences through solidarity suggests that, for a between-subjects experimental design, a sample of approximately 400–450 participants per condition is sufficient to detect small but meaningful effects (d = .20) with 80% power (Pérez, Vicuña, & Ramos, 2024). We used this estimate as a conservative benchmark to ensure adequate power to detect the first-stage effect of shared discrimination on solidarity within each racial group, thereby increasing our sensitivity to detect potential group differences in the full indirect pathway. In partnership with YouGov, a leading online survey firm, we recruited 2,804 Asian, Black, and Hispanic/Latino individuals living in the U.S. between October 9 and October 25, 2024. These respondents were matched to a nationally representative sampling frame based on gender, age, race, and education using census data (e.g., the American Community Survey), which yielded a final sample of 850 Asian, 850 Black, and 850 Latino adults. Key demographic information for each racial group is listed below in Table 1.

Table 1

Key Demographic Information

	Black	Latino	Asian
Age (years)	45.76	43.97	46.06
Gender (%)
Man	45.53	46.24	42.24
Woman	53.29	52.94	47.29
Non-binary	0.24	0.82	1.18
College educated (%)	56.23	47.18	79.65
Liberal ideology (1-5)	3.35	3.06	3.26
Partisanship (%)
Democrat	60.00	40.82	39.76
Republican	11.41	21.76	15.88
Independent	16.71	22.24	33.18
Registered Voters (%)	84.24	79.06	78.71
Foreign-born (%)	13.29	18.71	46.47

Experimental Manipulation

Prior studies testing the indirect effect of shared discrimination on policy preferences have leveraged Zou and Cheryan’s (2017) axes of discrimination framework to heighten perceptions of similarity between one racial minority ingroup and one racial minority outgroup at a time. For example, articles presented to Black Americans emphasized the shared second-class citizenship status experienced by Latinos. We extended this design by seeking to evoke a sense of shared discrimination between participants’ racial ingroup and two minority outgroups simultaneously. Specifically, participants in the treatment condition read an article describing rising hate crimes toward both of the other racial minority groups: Black participants read about hate crimes against Latinos and Asians; Latino participants read about hate crimes against Blacks and Asians; and Asian participants read about hate crimes against Blacks and Latinos.

The treatment article, titled “Never Fully American, Always an Outsider: 2024 Highlights the Decades-Long Exclusion of [Asian and Latino/Black and Asian/Black and Latino] People in the U.S.”, underscored both the foreignness and inferiority axes of discrimination. The article detailed recent increases in hate crimes, accompanied by two graphs (one for each racial outgroup) depicting a rise in reported incidents from approximately 200 in 2023 to about 500 in 2024. The article concluded with the following passage, explicitly linking the discrimination faced by the two racial minority outgroups to participants’ own racial group:

As a recent victim of one of these hate crimes toward [Asian and Latino / Black and Asian/Black and Latino] people stated (on condition of anonymity): “It’s so scary—and frustrating—that you can give so much of yourself to this country, and yet still be treated like an outsider, even if you and your family have been here for generations.” These words ring true among many [African/Latino/Asian] Americans throughout the U.S., who have been the targets of discrimination for decades.

The control article, comparable in length and style to the treatment article and slightly modified from prior studies, discussed the extinction of giant tortoises—a neutral, apolitical topic. The full-length articles are available in section SM.1 of the Supplementary Materials. (https://osf.io/f2rau).

Procedure

Participants from the YouGov panel who consented to take part in the study were directed to an online survey that began with a series of demographic questions, including education, gender, and partisanship. Following an attention check,^[1] participants were randomly assigned to either the control or treatment condition. After reading the assigned article, they completed a manipulation check^[2] before answering three items measuring solidarity with PoC. These items, previously validated through confirmatory factor analyses (Pérez, Goldman, et al., 2024), demonstrated good reliability within each racial group (Black = .68; Latino = .68; Asian = .79). They were rated on a 5-point scale from 1 (strongly disagree) to 5 (strongly agree) and included: “I feel solidarity with people of color, which includes Asian, Black, and Latino people,” “The problems of Black, Latino, Asian, and other people of color are similar enough for them to be allies,” and “What happens to people of color in this country has something to do with what happens in my life as a [Black / Latino / Asian] person.” The final item was matched to the race of respondents.

Participants next completed three items assessing their intention to vote for a candidate who would represent the best interests of PoC (Black = .89; Latino = .90; Asian = .93). Using the same 1 (strongly disagree) to 5 (strongly agree) scale, participants rated their agreement with statements such as “I am ready to vote for a presidential candidate who cares about the interests of Asian Americans, Black Americans, Latinos, and other people of color,” “I intend to vote for a presidential candidate that really listens to the concerns of Asian Americans, Black Americans, Latinos, and other people of color,” and “I plan to vote for a presidential candidate that has in mind the well-being of Asian Americans, Black Americans, Latinos, and other people of color.” Lastly, participants indicated their intended vote for Donald Trump (Republican), Kamala Harris (Democrat), another candidate, or selected “I do not intend to vote” in the upcoming 2024 presidential election. Those who chose not to vote were asked a follow-up question identifying their preferred candidate if they were to vote. To retain maximum statistical power, candidate preferences were then aggregated across both voters and non-voters to create a composite variable representing intent to vote for Harris. This variable was dummy coded as 0 (vote for Trump or another candidate) and 1 (vote for Harris) for subsequent analyses.

Results

We first conducted a 2 (Condition: Control, Shared Discrimination) × 3 (Race: Black, Latino, Asian) between-subjects ANOVA using IBM SPSS Statistics (v.30.0) to examine whether solidarity, intent to vote for a PoC representative, and intent to vote for Harris differed by experimental condition and participant race (Table 2). This approach provides a direct test of our manipulation’s ability to elicit solidarity and strengthen voting intentions, as well as its potential moderation by participant race. Consistent with prior meta-analytic research (Pérez, Vicuña, & Ramos, 2024), we found a significant main effect of condition on our proposed mediator such that participants who were exposed to shared discrimination appeals reported greater solidarity with PoC than participants in the control condition. However, contrary to our hypothesis, we found a non-significant interaction between race and condition, suggesting that shared discrimination motivates solidarity uniformly across groups (Table 3).

Next, we examined whether shared discrimination directly influenced participants’ voting intentions. We found no significant differences in participants’ intent to vote for a PoC representative or intent to vote for Harris by experimental condition. To assess whether these null effects reflect true absence of influence, we conducted two one-sided tests for each of our outcomes (TOST; Rainey, 2014). Drawing on prior research on voter turnout (Gerber & Green, 2000; Green & Gerber, 2019), we defined any effect smaller than ±2 percentage points in intent to vote for Harris as substantively negligible, corresponding to a ±0.08 change on our intent to vote for a PoC representative scale. The observed difference in intent to vote for a PoC representative between treatment and control conditions was 0.057 with 90% CI [–0.004, 0.118]. Similarly, the difference in proportion of participants intending to vote for Harris was 0.001 with 90% CI [–0.030, 0.032]. These intervals indicate that exposure to shared discrimination reminders could increase intent to vote for a PoC representative by as much as 0.12 points on our raw scale, t(2547) = -0.62, p = .268, or cause a 3.2 percentage point increase in intent to turnout for Harris, z = -0.93, p = .177. Thus, we cannot fully reject the possibility of substantively meaningful direct effects of shared discrimination on voting intentions.

Table 2

Mean (M), Standard Deviation (SD), and 95% Confidence Interval (CI) of Intent to Vote for a PoC Representative and Kamala Harris by Participant Race and Treatment Condition

		Control			Shared Discrimination
		M	SD	95% CI	M	SD		95% CI
Solidarity with PoC	Black	3.67	0.89	[3.54, 3.75]	3.96	0.85	[3.88, 4.04]
	Latino	3.51	0.83	[3.44, 3.59]	3.64	0.88	[3.56, 3.73]
	Asian	3.45	0.95	[3.36, 3.54]	3.57	0.94	[3.48, 3.66]
Vote for PoC Representative	Black	4.28	0.91	[4.20, 4.37]	4.36	0.88	[4.27, 4.44]
	Latino	4.02	0.91	[3.93, 4.11]	4.10	0.98	[4.01, 4.19]
	Asian	4.11	0.94	[4.02, 4.20]	4.13	0.94	[4.04, 4.22]
Vote for Harris	Black	.73	0.45	[.68, .77]	.76	0.43	[.72, .81]
	Latino	.58	0.49	[.53, .62]	.54	0.50	[.50, .59]
	Asian	.65	0.48	[.61, .70]	.66	0.48	[.61, .70]

Table 3

Main and Interaction Effects of Race and Shared Discrimination on Solidarity, Intent to Vote for a PoC Representative, and Intent to Vote for Harris.

		Race	Shared Discrimination	Race x Shared Discrimination
	df	2	1	2
Solidarity with PoC	F(df, 2542)	28.14	26.82	2.67
	p	<.001	<.001	.070
	η²ₚ	.010	.022	.002
Vote for a PoC Representative	F(df, 2540)	17.90	2.38	.19
	p	<.001	.123	.828
	η²ₚ	.014	.001	.000
Vote for Harris	F(df, 2541)	32.81	.01	1.35
	p	<.001	.930	.260
	η²ₚ	.025	.000	.001

To extend on prior mediation models demonstrating that shared discrimination increases policy support through solidarity (Pérez, Vicuña, & Ramos, 2024), we simultaneously estimate shared discrimination’s indirect effect on voting intentions and test for racial moderation of the solidarity–voting intention pathways using multi-group structural equation modeling (SEM) in Mplus 8.7. The model included both observed and latent constructs and was estimated using the weighted least squares mean and variance adjusted (WLSMV) estimator, which is robust to non-normality and appropriate for models including both categorical (voting for Harris) and continuous (voting for a PoC representative) outcomes (Li, 2021). Guided by the nonsignificant treatment by race interactions observed in the ANOVA, we constrained the direct paths from shared discrimination to solidarity and to both voting intention outcomes to be equal across racial groups^[3] (Figure 1). The model met conventional benchmarks for good fit, χ²(70) = 113.02, p < .001, CFI = .986, RMSEA = .027, SRMR = .034, which allows for a closer examination of group-specific standardized path coefficients from solidarity to vote intentions (Hu & Bentler, 1999).

Racial Moderation of Solidarity’s Effect on Vote Intentions

In line with prior work positioning Black Americans as the most prototypical PoC, we hypothesized that solidarity would be more positively associated with voting for a PoC representative among Black Americans than among Asian and Latino Americans. Contrary to this expectation, we found no evidence that the effect of solidarity on voting for a PoC representative was moderated by race: the differences between Black and Latino Americans (Δβ = -.02, SE = .06, 95% CI [-.14, .10], p = .749) and between Black and Asian Americans (Δβ = -.05, SE = .05, 95% CI [-.15, .06],  p = .369) were not statistically significant. We also hypothesized that the relationship between PoC solidarity and voting for Harris would vary by race, such that it would be more strongly associated with electoral support among Asian and Latino Americans relative to Black Americans, who are traditionally strong Democratic supporters. Consistent with this hypothesis, the association between solidarity and voting for Harris was significantly weaker among Black participants than among Latino (Δβ = -.26, SE = .07, 95% CI [-.40, -.12], p < .001) and Asian participants (Δβ = -.23, SE = .07, 95% CI [-.36, -.10], p < .001). No significant difference in effect size emerged between Latino and Asian participants (Δβ = .03, SE = .06, 95% CI [-.09, .15], p = .584).

In our theorizing, we posited that greater solidarity with PoC would motivate voting for Harris because of her affiliation with the Democratic Party—a party widely perceived as supportive of PoC interests. If this reasoning were correct, the two outcomes—intentions to vote for a PoC representative and to vote for Harris—should be correlated, and solidarity should exert a similarly directed effect on both. Although the outcomes were significantly but weakly correlated across all three racial groups (r < .30), solidarity was associated with significantly greater intent to vote for a PoC representative than for Harris among Black (Δβ = .31, SE = .07, 95% CI [.17, .44], p < .001) and Asian (Δβ = .13, SE = .05, 95% CI [.03, .23], p = .013) participants. In contrast, among Latino participants, solidarity was associated with comparable intent to vote for a PoC representative and for Harris (Δβ = .07, SE = .06, 95% CI [-.05, .19], p = .276).

Figure 1

Standardized Estimates from a Partially Constrained SEM Predicting Intent to Vote for a PoC Representative and Intent to Vote for Harris.

Shared Discrimination’s Indirect Effect on Voting Intentions

In line with prior research, the shared discrimination manipulation showed a positive indirect association with voting intentions through increased solidarity with PoC across all three racial groups^[4] (Table 4). However, these indirect effects did not differ meaningfully across groups. The standardized indirect effects (IE) were statistically indistinguishable between Black and Latino (ΔIE = -.003, SE = .021, 95% CI [-.046, .040], p = .891), Black and Asian (ΔIE = .006, SE = .020, 95% CI [-.033, .045], p = .762), and Latino and Asian (ΔIE = .009, SE = .021, 95% CI [-.031, .049], p = .662) participants for intentions to vote for a PoC representative. The same pattern held for voting intentions toward Harris: no significant differences emerged between Black and Asian (ΔIE = -.019, SE = .015, 95% CI [-.048, .010], p = .201), or Latino and Asian participants (ΔIE = .015, SE = .018, 95% CI [-.020, .050], p = .400). The only exception was a modestly stronger indirect effect for Latinos relative to Black participants (ΔIE = -.034, SE = .017, 95% CI [-.068, -.0003], p = .048). All results presented thus far were robust to controls for age, gender, education, and nativity (SM.3) and replicated among registered voters (SM.4) and among participants who passed the manipulation check (SM.5).

Considering the possibility of confounding—particularly in cross-sectional data, where the association between solidarity and vote intention may be influenced by an unmeasured third variable (Rohrer & Arslan, 2021)—we conducted sensitivity analyses in Stata to assess the robustness of our findings. This analysis produced a sensitivity parameter (ρ) for the observed association between solidarity and voting for Harris across our three studies, estimating the degree to which an omitted confounder would need to correlate with both the mediator and the outcome to reduce the indirect effect to zero (Imai & Yamamoto, 2013). As shown in Table 5, only a relatively strong confounder (ρ ~ .421) could nullify the significant positive association between solidarity and voting for Harris. While this analysis does not establish causality, it supports the plausibility of the proposed mechanism and underscores the need for future experimental work to further evaluate this pathway.

Table 4

Standardized Indirect Effects (IE) and Standard Errors (SE) from Shared Discrimination to Vote Intentions

	PoC Representative			Harris vs. Trump/Other			PoC Representative – Harris
	IE (SE)	95% CI	p	IE (SE)	95% CI	p	ΔIE (SE)	95% CI	p
Black	.07 (.02)	[.04, .10]	<.001	.03 (.01)	[.02, .06]	.001	.04 (.02)	[.004, .07]	.031
Latino	.07 (.02)	[.05, .11]	<.001	.07 (.01)	[.04, .10]	<.001	.01 (.02)	[-.02, .05]	.411
Asian	.07 (.01)	[.04, .09]	<.001	.05 (.01)	[.03, .08]	<.001	.01 (.02)	[-.03, .05]	.707

Table 5

Sensitivity Parameter for the Correlation between Solidarity and Voting for Harris

	Voting for Harris
	Black	Latino	Asian
Solidarity (ρ)	.408	.402	.454

Discussion

In the run-up to the 2024 U.S. presidential election, we tested whether making shared experiences of discrimination salient would heighten solidarity with PoC which, in turn, would be positively associated with voting intentions. We also examined whether these effects were moderated by racial group membership using nationally representative samples of Black, Latino, and Asian Americans. Across all three groups, shared discrimination appeals produced a comparable increase in solidarity with PoC but had no direct influence on voting intentions. When solidarity was included as a mediator, however, a different pattern emerged. The resulting increase in solidarity was positively associated with both intent to vote for a PoC representative and intent to vote for Kamala Harris—outcomes that are moderately correlated with actual voting behavior (Ajzen, 1985; Randall & Wolff, 1994). At the same time, the direct path from shared discrimination to voting intentions became small and negative (Figure 1), consistent with a suppressor effect^[5] (MacKinnon et al., 2000). This pattern suggests that shared discrimination appeals may simultaneously activate competing processes: one that mobilizes solidarity, and another that evokes threat– consistent with the literature on interminority solidarity (Craig & Richeson, 2012, 2014, 2016).

Although the strength of the association between solidarity and intent to vote for a PoC representative did not differ across racial groups, the relationship between solidarity and intent to vote for Harris was significantly stronger among Asian and Latino Americans than among Black Americans. This pattern likely reflects a political ceiling effect, as Black Americans already display consistently high baseline support for Democratic candidates (Hartig et al., 2025; White & Laird, 2020). By contrast, Asian and Latino Americans showed greater variability in their electoral responses to solidarity cues, suggesting that solidarity with PoC may serve as a more potent mobilizing force when partisan loyalties are less firmly established. Altogether, our findings indicate that shared discrimination appeals can indirectly shape voting intentions by strengthening solidarity among racially minoritized groups, even as countervailing processes constrain their direct influence.

At first glance, these findings might appear inconsistent with media coverage of the 2024 election results, which highlighted the non-trivial support that Donald Trump received among voters of color (Brown et al., 2024; Dowd, 2024; Frey, 2024; Montanaro et al., 2025). Given that Donald Trump narrowly edged out Kamala Harris in the popular vote, the indirect associations observed in our study may not have fully translated to the real-world electoral context. This apparent disconnect likely reflects the situational dependence of identity salience (Turner et al., 1987). The self-conceptions and electoral choices of voters hinge on the identities and narratives made salient during campaigns (Bejarano et al., 2021; Sides et al., 2022; Vavreck, 2001). In 2024, however, the Harris campaign deliberately deemphasized race and identity politics, positioning her as Donald Trump’s opposite on economic issues rather than as a candidate representing racially minoritized communities (Keith, 2024).

Despite the historic potential of being the first Black, Indian, and female president, Harris downplayed her multiracial identity, stating, “I am running because I believe that I am the best person to do this job at this moment for all Americans, regardless of race and gender” (Browning, 2024; Keith, 2024). Consequently, unlike in our experimental design—which employed a facilitative design (Sniderman, 2011) providing participants with a compelling reason (shared discrimination) to act in line with existing predispositions (solidarity with other racially minoritized groups)—voters in the real election were not presented with similar cues linking their group’s experiences to a broader PoC coalition. This strategic choice likely constrained Harris’s ability to act as an “identity entrepreneur” (Reicher & Hopkins, 1996; Simon & Klandermans, 2001)—a leader who mobilizes support by constructing and activating a shared collective identity. Such leaders define a group’s common struggles, allies, and adversaries, thereby shaping a politicized sense of “us” that underpins solidarity and collective action (Hogg & Reid, 2006; Huddy & Bankert, 2017).

In prior elections, candidates like Barack Obama effectively fulfilled this role, strengthening Black identification with the Democratic Party and broadening coalitions among people of color (Huddy & Bankert, 2017). In contrast, Harris’s avoidance of identity-based appeals meant that she was not perceived as a clear PoC representative, especially among Black voters—for whom shared discrimination was more strongly associated with the intent to vote for a candidate perceived to represent PoC interests than the intent to vote for Harris through solidarity (Table 4). By minimizing “identity politics,” Harris sought to broaden her appeal among White moderates. Yet in doing so, she relinquished the opportunity to clearly position herself as the prototypical representative of voters of color—an identity our results suggest would have strengthened her support base. Meanwhile, Trump’s overt “us versus them” rhetoric likely heightened identity boundaries among his supporters, amplifying asymmetries in mobilization. Together, these findings highlight the conditional nature of solidarity’s political impact: its capacity to shape electoral behavior depends not only on shared experiences of discrimination but also on whether political elites activate and channel those shared identities into collective political purpose.

Our study did not detect significant direct effects of shared discrimination on voting intentions. While this might suggest that the influence of shared discrimination does not extend to more consequential political behaviors such as voting, equivalence tests indicate that we cannot rule out substantively meaningful effects. This finding aligns with meta-analytic evidence showing that, although the overall effect of shared discrimination on solidarity and outgroup policy support is positive and statistically significant, individual cross-sectional studies vary in both direction and magnitude of effects (Pérez, Vicuña, & Ramos, 2024). Across the five cross-sectional experiments included in the meta-analysis, estimated effects ranged from d = 0.012 to 0.362 and p-values ranged from 0.001 to 0.898. Thus, further investigation of these direct pathways is warranted.

Although our study did produce significant direct effects of shared discrimination on solidarity, the magnitude of this effect should be interpreted with caution. Participants in the control group read a neutral article about turtles which—despite its use in prior conceptually similar studies (Pérez, Vicuña, & Ramos, 2024)—was also devoid of broader social, political, or self-relevant content. This mismatch in topic relevance raises the possibility that the observed effects partly reflect increased salience of race rather than the specific framing of shared discrimination per se. Future research could provide a more stringent test of these effects by using a more closely matched control stimulus, such as the one employed by Craig and Richeson (2012), in which race was made salient for Black and Latino participants through an ostensibly unrelated article describing higher rates and severity of lupus in their respective racial groups. Such a design would allow for a stronger assessment of whether intergroup solidarity arises uniquely from recognizing shared discrimination or more generally from reflecting on one’s racial identity.

Beyond the content of the control condition, the type of discrimination emphasized in our manipulation also warrants consideration. Whereas previous studies have often made shared discrimination salient through descriptions of everyday or historical injustices (e.g., Cortland et al., 2017; Craig & Richeson, 2016), the present study focused on hate crimes—acute and highly visible manifestations of racialized discrimination. We selected this approach to highlight a form of discrimination that would be salient and recognizable across all three racial groups. In light of the recent surge in hate crimes targeting Asian Americans during COVID-19 (Lantz et al., 2022; Lim et al., 2022; Reny & Barreto, 2020; Ruiz et al., 2023), ongoing discussions of police violence toward Black Americans (Toosi et al., 2021; Toraif et al., 2023; Wilson, 2025), and the resurgence of anti-immigrant rhetoric that sets the stage for hate crimes against Latinos (Hamedy, 2018; Kang & Yang, 2021; Stacey et al., 2011), this interpersonal yet acute form of discrimination was expected to resonate broadly. Nevertheless, we acknowledge that this focus diverges from prior work emphasizing chronic or structural forms of discrimination, and we note that future research should examine whether similar effects emerge when shared discrimination is made salient through everyday or historical examples.

Finally, as with most survey experiments, our measures relied on self-reported attitudes and behavioral intentions. Although voting intentions are among the strongest predictors of actual turnout (Ajzen, 1985; Randall & Wolff, 1994), they remain imperfect proxies for real-world behavior. Future research could incorporate longitudinal follow-ups and validated behavioral outcomes (e.g., voter records or political donations) to assess whether shared discrimination appeals translate into sustained political engagement.

Conflicts of Interest

The authors declare no competing interests.

Acknowledgements

This project was funded by the Russell Sage Foundation.

Data Availability Statement

Data, code, and study materials are stored in OSF: https://osf.io/f2rau/

Supplementary Materials

The supplementary materials can be found here.

Author Contributions

E.P. and K.R. designed the study. K.R. analyzed the data and drafted the introduction, methods, results, and discussion sections. Both authors provided critical revisions.

Endnotes

1. All participants in our dataset passed the attention check, which required them to select the “oppose” option on a scale ranging from 1 (strongly oppose) to 7 (strongly support).All participants in our dataset passed the attention check, which required them to select the “oppose” option on a scale ranging from 1 (All participants in our dataset passed the attention check, which required them to select the “oppose” option on a scale ranging from 1 (strongly opposestrongly oppose) to 7 () to 7 (strongly supportstrongly support).).
2. Participants responded to a true or false statement matched to their assigned treatment condition: “The information I read highlighted how giant tortoises are in decline” or “The information I read highlighted how [respondent’s race] people are still viewed as outsiders and not fully American.” Most Black (86.0%), Latino (83.8%), and Asian (89.4%) participants passed this manipulation check. To retain maximum statistical power, participants who failed the check were included in all main analyses. Results from analyses restricted to participants who passed the check are reported in SM.5. Participants responded to a true or false statement matched to their assigned treatment condition: “The information I read highlighted how giant tortoises are in decline” or “The information I read highlighted how [respondent’s race] people are still viewed as outsiders and not fully American.” Most Black (86.0%), Latino (83.8%), and Asian (89.4%) participants passed this manipulation check. To retain maximum statistical power, participants who failed the check were included in all main analyses. Results from analyses restricted to participants who passed the check are reported in SM.5. Participants responded to a true or false statement matched to their assigned treatment condition: “The information I read highlighted how giant tortoises are in decline” or “The information I read highlighted how [respondent’s race] people are still viewed as outsiders and not fully American.” Most Black (86.0%), Latino (83.8%), and Asian (89.4%) participants passed this manipulation check. To retain maximum statistical power, participants who failed the check were included in all main analyses. Results from analyses restricted to participants who passed the check are reported in SM.5.
3. We compared this partially constrained model to both an unconstrained model and a fully constrained model. Model comparison tests indicated that the partially constrained model provided the best balance of parsimony and fit (see SM.2).We compared this partially constrained model to both an unconstrained model and a fully constrained model. Model comparison tests indicated that the partially constrained model provided the best balance of parsimony and fit (see SM.2).We compared this partially constrained model to both an unconstrained model and a fully constrained model. Model comparison tests indicated that the partially constrained model provided the best balance of parsimony and fit (see SM.2).
4. We note that the indirect effect of shared discrimination on voting intentions is imprecisely estimated (p = .07) among Asian participants when all structural paths are freely estimated (SM.2). We note that the indirect effect of shared discrimination on voting intentions is imprecisely estimated ( We note that the indirect effect of shared discrimination on voting intentions is imprecisely estimated (pp = .07) among Asian participants = .07) among Asian participants when all structural paths are freely estimated when all structural paths are freely estimated (SM.2).(SM.2).
5. In the unconstrained model, we find that the negative direct effect of shared discrimination on intent to vote for Harris is significant in the Latino sample. We discuss this result in the context of the 2024 election in SM.2. In the unconstrained model, we find that the negative direct effect of shared discrimination on intent to vote for Harris is significant in the Latino sample. We discuss this result in the context of the 2024 election in SM.2. In the unconstrained model, we find that the negative direct effect of shared discrimination on intent to vote for Harris is significant in the Latino sample. We discuss this result in the context of the 2024 election in SM.2.

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