The irony of (romantic) harmony: Heterosexual romantic relationships can drive women’s justification of the gender hierarchy

Abstract

Even though gender inequality is evident across life domains, women often justify the gender hierarchy. We examined whether the very closeness that heterosexual women share with their male romantic partners predicts their justification of gender inequality. We drew on intergroup-related research, showing that positive perceptions that minority groups develop within harmonious intergroup interactions, generalize to affect their views of group-based inequality. We expected that to the extent that women experience their romantic relationships positively, they will be more accepting of gender inequality within their homes, and these perceptions will generalize to predict justification of macro-level gender inequality. Five correlational and two experimental studies supported this prediction. The more women rated (or were primed with) their relationship as positive, the more they justified the gender social system. This was mediated by women’s perception of their housework division as fair, and was less pronounced among feminists. Implications regarding social change are discussed.

Keywords

gender inequality intergroup contact system justification

Gender disparities favoring men are evident across numerous and diverse fields. Women are extremely underrepresented in leadership positions in politics (The World Bank, 2017), top companies (Dezső et al., 2016), and in academia (European Commission, 2016). Women are also far less likely to own assets and land (Yokying & Lambrecht, 2019), and they earn less money than men across countries (Catalyst, 2022). Gender inequality is further manifested within people’s homes. Women, worldwide, tend to do a larger share of housework than their male partners, regardless of their age, income, and workloads (International Labour Organization, 2016; Treas & Tai, 2016). Given this widespread and stable inequality, one would expect women to object to it. However, a recent report (United Nations Development Programme, 2020) reveals that most men and women accept social norms which proscribe women’s rights. In the present research, we investigated processes that may intensify women’s tendency to accept and justify the social system which disadvantages them. Specifically, we examined whether heterosexual romantic relationships between men and women predict women’s justification of gender inequality.

Considerable evidence suggests that women, often, do not acknowledge the gender-based hierarchy. For example, relative to other disadvantaged groups (specifically Black and Latino Americans), women are least likely to perceive themselves as discriminated against on the basis of their group membership (Harnois, 2015). Consistent with this, 54% of U.S. women perceive no difference between men and women in how easy life is for them, with an additional 5% indicating that women have easier lives than men (Pew Research Center, 2017). Other findings indicate that women, under circumstances that underscore beliefs in individual merit, blame gender-based rejection on themselves, and minimize attribution to gender discrimination (McCoy & Major, 2007). Moreover, when women do acknowledge the existence of gender inequality, they are readily inclined to accept it and view it as justified (Glick et al., 2000).

The tendency to justify gender hierarchy on part of women can be attributed to people’s basic motivation to accept and justify existing social arrangements, which are often unequal (Jost & Banaji, 1994). Counterintuitively, even members of disadvantaged groups are motivated to justify the unequal system, in order to reduce uncertainty and threat (Jost et al., 2003). When it comes to gender, justifying the current system reflects tolerance for gender-specific inequality and is termed “gender system justification” (Jost & Kay, 2005). As with other psychological motives, gender system justification can be triggered, or undermined, depending on the social context (see Jost, 2019). For example, exposure to prevalent stereotypes of women as communal was found to increase women’s gender system justification (Jost & Kay, 2005). Similarly, exposure to benevolent sexism, a view that idealizes women and considers them in need of men’s protection (Glick & Fiske, 2001), also increased perception of the gender system as fair (Becker & Wright, 2011). This can be a result of failure to recognize benevolently sexist acts (e.g., overly protecting women) as sexism, and instead considering them as a positive element in gender relations (Barreto & Ellemers, 2005). Even when these behaviors are identified as a form of sexism, their warm and affectionate content leads women to view benevolent sexist men as lower in hostile sexism and more supportive of gender equality (Hopkins-Doyle et al., 2019).

In the current research, we propose an additional factor that may increase women’s tendency to justify the gender social system. Gender inequality is a unique case of group-based hierarchy: it involves, on the one hand, pervasive structural disparities, and on the other hand, countless instances of interpersonal closeness between the advantaged and disadvantaged group in the form of heterosexual romantic relationships. Our approach links the two by proposing that processes that occur within heterosexual romantic relationships can feed macro-level hierarchy by driving psychological processes of justification on part of women.

To develop our reasoning, we consider heterosexual romantic relationships as an intergroup encounter involving partners that belong to distinct social groups characterized by a stable and widespread inequality (Jackman, 1994; Pratto, 1996). From this standpoint, psychological processes that take place in asymmetrical intergroup relations, are relevant for understanding romantic relationships and their consequences. Specifically, we propose that heterosexual romantic relationships can be analyzed from the perspective of intergroup contact, which provides insights about processes that disadvantaged group members go through whilst in positive encounters with members of the advantaged group (see Dixon et al., 2012; Saguy, 2018). Relying on research conducted among ethnic and sexual minorities, we suggest that to the extent that women experience their heterosexual romantic relationships as positive, they will tend to justify macro-level gender hierarchy. This will be the case because women in positive romantic relationships will tend to view inequality with their male partners as justified. Such perceptions can, in turn, be generalized to shape perceptions about macro-level gender relations.

Romantic Relationship as Optimal Intergroup Contact

Research involving members of ethnic and/or racial groups shows that when members of opposing groups come together in a meaningful encounter, intergroup attitudes improve and stereotyping and discrimination decrease (Pettigrew, 1998; Pettigrew & Tropp, 2006). This improvement is considered to be more likely when the encounters are “optimal” or positive, that is, contain Allport’s (1954) conditions for optimal contact—common goals, intimacy, equal status, and institutional support. Even though there is a lack of experimental support for the necessity of these conditions (Paluck et al., 2019), the notion of optimal contact has inspired a wealth of research and desegregation policies, which often involve one or more of the specified conditions (Pettigrew, 1998; Pettigrew & Tropp, 2006).

Heterosexual romantic relationships usually involve the characteristics of positive or optimal contact, particularly in Western societies. Relationship partners often share mutual goals (e.g., successful child rearing); intimacy between the partners is integral to romantic relationships, and even though hierarchical relations between the groups are evident, romantic partners typically enter their relationship as equals (at least at a stated level). Moreover, social institutions, norms, and prevailing ideologies—all support committed romantic relationships (Day et al., 2011). Given the parallels between positive, or optimal, intergroup contact and romantic relationships, the outcomes of intergroup contact can also apply to heterosexual romantic relationships. Such outcomes go beyond reduction in prejudice and improvement in intergroup attitudes.

Indeed, intergroup contact was found to shape disadvantaged group members’ support for the status quo (for reviews, see Dixon et al., 2012; Saguy, 2018). Findings based on samples from 69 countries with 4883 members of ethnic and sexual minorities, revealed that having more and better contact with the respective majority group predicted less motivation to advance change in the status hierarchy (Hässler et al., 2020). These ideas are corroborated by experimental evidence. When a positive relationship between a disadvantaged and an advantaged group member is primed or formed, disadvantaged group members tend to be less concerned with group-based inequality (Saguy et al., 2009), and less motivated to support social change (Becker et al., 2013). For example, students from a low-status college interacted with a student from a higher status university under conditions that either mapped onto optimal contact (i.e., focused on commonalities under equal status conditions) or not (i.e., focused on differences between the schools). Participants in an additional control group had no experience of contact. Those who were part of an optimal encounter (relative to participants in both control groups) rated the macro-level inequality in academic institutions (e.g., having high status institutions get more governmental benefits, and as a result more opportunities for its graduates) as less troubling (Saguy & Chernyak-Hai, 2012; Study 1).

What might explain these effects? Close relationships often promote a focus on commonalities and distract attention away from group-based differences. This blurring of intergroup boundaries can weaken disadvantaged group members’ attention to group-based injustice (Wright & Lubensky, 2009; Saguy, 2018). Indeed, disadvantaged group members often perceive their advantaged group friends as fair and trustworthy (Saguy et al., 2009), and these views can generalize to affect perceptions regarding the outgroup as a whole (Pettigrew & Tropp, 2006, 2011), and the status quo (Saguy & Chernyak-Hai, 2012; Sengupta & Sibley, 2013). For example, the more they had contact with Whites, Black South Africans perceived less discrimination against their group (Dixon et al., 2010), and less ingroup deprivation (Cakal et al., 2011). Similarly, positive contact with White Americans decreased ethnic activism among Latino Americans and African Americans over time, and this was mediated by lowered perceptions of racial discrimination (Tropp et al., 2012). Thus, positive perceptions that develop within cross-group interpersonal relationships (e.g., a sense of trust in the advantaged outgroup friend), seem to extend and affect the way the macro-level hierarchy is viewed.

We propose that similar processes of generalization can occur in the context of heterosexual romantic relationships: to the extent that a romantic relationship with a man is experienced positively, a woman can come to view inequality within the relationship as just and fair, views that can generalize and affect larger perceptions of gender relations. Therefore, we are predicting a two-stage process: one concerns perceptions regarding the relationship itself, and the consequent one concerns the generalization of such perceptions to broader views of gender hierarchy.

Recent work lends support for this idea. Vázquez et al. (2020) showed that women who report having more positive contact with men, had more positive attitudes towards men (in general) and were less motivated to engage in collective action in support of women’s rights. This research, however, did not distinguish between general contact experiences that women have with men in multiple occasions (e.g., in one’s work, in one’s neighborhood, in one’s varied social circles, in one’s extended family) and contact that is part of a romantic relationship. We believe such distinction is important due to the unique features of romantic relationships that render them a special case of gender-based contact. Indeed, our hypothesized mediator is the justification of inequality within one’s own home. This mediator is rooted in the joint household element that is a common feature of romantic relationships and has little to do with general forms of contact. Thus, the current work diverges from that of Vázquez et al. (2020) in the specific focus on romantic contact and in the proposed underlining psychological mechanism.

Division of Housework as a Domain of Domestic Inequality

To assess perceptions of fairness within the relationship, we focused on women’s perceptions regarding the division of housework in their homes. By housework we refer to the routine tasks that are ongoing and essential for maintaining the family needs (e.g., laundry, cooking, childcare; Badr & Acitelli, 2008). Despite women’s increased presence in the paid workforce, and men’s increased contribution to the household chores, the division of household labor remains largely unequal, with women still doing the lion’s share of the work (International Labour Organization, 2016). This widespread disparity manifests across countries (Neilson & Stanfors, 2014), even in the most egalitarian ones (e.g., Norway and Denmark; Treas & Tai, 2016), and within households that include women with a high level of education who earn high wages and spend a similar amount of time doing paid work as their spouses (Horne et al., 2018; Young et al., 2013).

Despite the presence of domestic inequality, women tend to perceive the unbalanced division of housework in their homes as relatively just (Jansen et al., 2016; van Hooff, 2011). Thus, we can assume that (1) in most cases the division of housework is unequal and disadvantages women, and (2) that this disparity is not necessarily seen as unfair by women. Hence, given that we aim to study variability in perceptions of fairness within-relationship inequality, the domain of housework division seems to be a suitable one.

The Irony of Romantic Harmony

Integrating the research reviewed above, we propose that heterosexual romantic relationships can have ironic effects on perceptions of gender relations. Specifically, we offer the following process: to the extent that a woman perceives her heterosexual romantic relationship as positive (we define “positive” along the dimensions of optimal contact; Allport, 1954; Pettigrew, 1998), she would tend to view the distribution of housework between her and her spouse as more fair, regardless of the actual (in)equality of their division. This can be a result of positivity that characterizes optimal contact (as reviewed above), and can also reflect a motivation to maintain relative harmony at home. Individuals in satisfying heterosexual relationships tend to view their partners and relationships in the best light possible (Lemay & Clark, 2015; Murray, 1999)—a positivity bias that assists motivated perceivers to maintain close bonds with their partners. We further propose that once the (often gendered) housework division is viewed as fair, such justification would generalize to affect more general views of gender hierarchy as justified. This idea draws on the literature reviewed above, linking perceptions of fairness within close friendships with outgroup members to wider perceptions regarding the outgroup as a whole, and the status quo.

We do note a potential boundary condition to this effect. Specifically, we expect the predicted process to be less applicable to disadvantaged group members who are likely to have relatively solid ideological disapproval of the hierarchy (van Breen et al., 2017). Literature on intergroup harmony and social change shows that harmony is less predictive of reduced social change tendencies among disadvantaged group members who are strongly aware of group-based inequality (see Hasan-Aslih et al., 2018). Applying this to the current context, we would expect women who identify as feminists to be highly aware of gender inequality, so that their views regarding gender relations might be less influenced by processes of contact. Therefore, we considered level of feminism as a boundary condition to the proposed process.

The Current Research

We conducted five correlational studies across different Western samples (Studies 1a–1e), and two experimental studies (Studies 2 and 3), which were held in the United States. In Study 1, across samples, we first examined the association between optimal contact within romantic relationships and justification of gender inequality. Second, we tested the role of perceptions regarding housework fairness as a mediator of this association. In the following studies, we experimentally tested the predicted process by breaking it down to its two main components: in Experiment 2, we tested the “a” path of the mediation model, namely, whether optimal contact (experimentally manipulated) predicted perceptions of one’s housework division as fair (regardless of how much work is actually performed). In Experiment 3, we tested the “b” path by directly manipulating the mediator, perceptions of fairness regarding the housework, and examined whether it predicted macro-level justification of gender hierarchy. This experiment was preregistered at AsPredicted.org (see relevant URL in the Method section of Study 3).

Participants across studies were heterosexual women who are, or were, involved in a romantic relationship. In order to assess the degree to which women perceived their heterosexual relationship as corresponding to the dimensions of optimal contact, we developed a measure based on Pettigrew’s (1998) optimal intergroup contact principles (assessing the extent to which the relationship is characterized by common goals, intimacy, equal status, and normative support). Gender related system justification (GSJ hereafter) served as our outcome measure across all studies (Jost & Kay, 2005). We considered women’s fairness perceptions regarding the division of housework in their households (perceived fairness of housework hereafter) as our mediator, which was measured in Studies 1 and 2, and manipulated in Study 3. We further assessed feminist identification across studies to test its potential moderating role.

We hypothesized that, to the extent that the romantic relationship will be rated more positively (as indicated by items assessing dimensions of optimal contact), or primed as (more or less) positive (Study 2), women would show stronger inclination to justify the gender social system. We further expected this association to be explained by perceptions of greater fairness of one’s own housework division. Thus, we predicted an indirect effect (optimal contact → perceived fairness of housework → gender system justification). To validate this indirect hypothesis, Study 3 directly manipulated the mediator (perceived fairness of housework) and tested its causal effect on gender system justification.

To verify that, consistent with much data (Horne et al., 2018; Treas & Lui, 2013; Treas & Tai, 2016), our female participants are those doing a greater share of housework relative to their partners, in two of the correlational studies (1d and 1e) and in the experiments (Studies 2 and 3) we assessed the reported amount of actual housework performed, and controlled for this across analyses.

Study 1(a–e)

Across five correlational studies in different Western samples of women, we tested the associations between optimal contact within heterosexual romantic relationships and GSJ, via the expected mediating role of perceived fairness of housework. To parse out variance due to demographics and relationship characteristics, across analyses we controlled for participants’ age, relationship status (whether one is reporting on a current or past relationship), and relationship duration. We also controlled for participants’ general tendency to support group-based hierarchy as assessed by the social dominance orientation scale (SDO; Pratto et al., 1994), given the association of SDO with gender-related attitudes (such as sexism and tolerance for sexual harassment). In the last two studies (1d and 1e), we also assessed the relative contribution of participants to household and childcare-related tasks, and controlled for these variables as well. All studies included a measure of feminism, in attempt to test for potential moderation effects.

Method

Sample size

In Studies 1a and 1b we aimed for a sample of 147 participants, based on a calculation conducted using the G*Power software (Faul et al., 2009). We considered a low-medium effect size (r = .20; estimated based on previous research on contact and system justification indices; Saguy & Chermyak-Hai, 2012), an alpha level of .05 (one-tailed), and 80% power. Given the inconsistency in the effect sizes obtained in Studies 1a and 1b, we increased the sample size in the rest of the studies.

Participants

Across studies, participants were women. We excluded respondents who did not identify as heterosexual in the sexual orientation item (in Study 1d exclusion was based on the partner’s gender), and participants who were never involved in a romantic relationship. We also excluded those who left these items blank. Detailed information regarding exclusions for each study is presented in the supplementary material online.

In four of our studies (all, except for Study 1d), participants were recruited through an online survey platform/panel, and took part in the study in exchange for monetary compensation of around 1.5 USD. Study 1a was conducted in the United States (via Amazon Mechanical Turk; N = 137), Study 1b was conducted in Israel (via Midgam Panel; N = 136), Study 1c was run in the United Kingdom (via Prolific; N = 198), Study 1d was run in Spain and involved students from a distant-learning University who volunteered to complete an online study (N = 229), and Study 1e was run again in the United States (via Amazon Mechanical Turk; N = 288). Means and standard deviations of demographics and personality variables for Studies 1a–1e are presented in Table 1.

Table 1.

Means and standard deviations for all variables in Studies 1a–1e.

Variable	M (SD)
Variable	Study 1aU.S.	Study 1bIsrael	Study 1cU.K.	Study 1d Spain	Study 1eU.S.
Optimal contact (1-7 scale)	5.61 (1.15)	5.51 (1.2)	5.59 (1.07)	5.24 (1.16)	5.69 (1.04)
GSJ (1-7 scale)	4.21 (1.28)	3.46 (1.02)	4.10 (0.88)	2.80 (0.77)	4.06 (1.15)
Perceived fairness of housework (1-7 scale)	5.50 (1.51)	4.57 (1.82)	5.13 (1.53)	4.51 (1.87)	5.07 (1.66)
Feminist identification (1-7 scale)	3.80 (1.93)	4.21 (1.46)	4.43 (1.70)	4.65 (1.63)	3.90 (1.94)
Age	34.67 (9.87)	43.93 (14.94)	37.28 (12.3)	37.10 (10.67)	35.78 (10.49)
Current relationship status (yes = in a relationship currently)	Yes 85.4% No 14.6%	Yes 78.7% No 21.3%	Yes 82.3% No 17.7%	Yes 73.4% No 26.6%	Yes 100%
Relationship duration (in years)	8.55 (8.68)	15.56 (13.27)	11.23 (10.41)	11.23 (10.41)	11.00 (9.98)
SDO (1-7 scale)	2.10 (1.24)	2.23 (0.95)	2.46 (1.10)	1.71 (0.84)	2.47 (1.23)
Household contribution (0%-100% in Study 1d; 1-5 in study 1e)				62.82 (21.29)	3.90 (9.95)
Childcare contribution (0%-100% in Study 1d; 1-5 in study 1e)				56.91 (31.48)	3.85 (0.90)

Measures

Unless otherwise specified, responses to all measures were given on a scale ranging from 1 = strongly disagree to 7 = strongly agree. Information regarding additional measured variables, which are not directly associated with the main hypotheses, is presented in the supplementary material.

Optimal contact

We assessed optimal contact with eight items developed for the purpose of this research. Each of the four conditions specified by Allport (1954), namely, common goals, intimacy, equal status, and normative support, was assessed with two corresponding items: “My partner and I share common goals”; “My goals are compatible with my partner’s goals”; “My partner and I know each other deeply”; “My partner is my best friend”; “My partner and I have equal status within our relationship”; “My partner and I treat each other as equals”; “Society encourages and supports relationships like my partner and I maintain”; “My social environment (parents, friends, relatives, etc.) is supportive of my relationship with my partner.” Given the strong correlations among items, which were designed to represent a single construct (i.e., optimal contact), they were averaged to create an optimal contact scale, α(Study 1a) = .89; α(1b) = .88; α(1c) = .89; α(1d) = .85; α(1e) = .91. The higher the score, the more participants perceive their romantic relationship positively (i.e., corresponds to the definition of optimal contact).

Gender System Justification (GSJ)

To assess justification of macro-level gender hierarchy, we used the validated 8 items scale based on Jost and Kay (2005) (e.g., “In general, relations between men and women are fair;” “Society is set up so that men and women usually get what they deserve”), α(1a) = .90; α(1b) = .76; α(1c) = .77; α(1d) = .68; α(1e) = .83. Reversed items were transformed, and the eight items were averaged into a GSJ scale with higher score reflecting greater justification of gender hierarchy.

Perceived fairness of housework

Participants indicated the degree to which they felt the division of household tasks between them and their partner (or previous significant partner) was justified, legitimate, and fair (based on Jost & Burgess, 2000), α(1a) = .97; α(1b) = .95; α(1c) = .96; α(1d) = .93; α(1e) = .96. We calculated a mean score of all items, so that a higher score on this scale reflects stronger fairness perception.

Feminist identification

Respondents were asked to indicate the extent to which they “consider themselves as feminists” and “identify with feminists,” on a scale ranging from 1 = not at all to 7 = very much so (adapted from Luhtanen & Crocker, 1992). The two items were averaged to a feminist identification measure, r(1a) = .81; r(1b) = .63; r(1c) = .73; r(1d) = .74; r(1e) = .89, ps < .001.

Social Dominance Orientation (SDO)

Four items from the 16-item SDO measure (Pratto et al., 2013) were used to examine participants’ support for group-based hierarchy (sample item: “Superior groups should dominate inferior groups”). Reversed items were transformed, and the four items were averaged to an SDO scale, so that a higher score reflects more support in group-based hierarchy, α(1a) = .86; α(1b) = .72; α(1c) = .80; α(1d) = .68; α(1e) = .80.

Household and childcare contribution

This measure was only included in the last two studies. Participants indicated, on two separate items, their relative contribution to household and childcare tasks (compared to their partner) on a scale ranging from 0% to a 100% (Study 1d) or from 1 = my partner does much more than I do to 5 = I do much more than my partner does (Study 1e).

Procedure

Participants were asked to complete a set of online questionnaires. In order to minimize problems of demand, the measures appeared in the following order: GSJ, details regarding the relationship (status, duration, relative contribution to housework, optimal contact, and perceived fairness of housework), SDO, additional demographic details, and feminist identification items.

Results

Means and standard deviations of the key variables are presented in Table 1. Levels of optimal contact were relatively high across studies (means are around 5.5 on a 1–7 scale), and levels of GSJ were around the midscale (means are around 4) for most samples, except for Study 1d, in which GSJ was lowest (M = 2.80). The relatively low levels of GSJ within the Spanish sample could reflect the relatively liberal Psychology student population (Vázquez & Lois, 2019). As further expected, Studies 1d and 1e revealed that women reported doing more housework than their male partners.

Zero-order relationships

Zero-order correlations between key variables are presented in Table 2. For presentation purposes, we display only the correlations of each variable with the independent variable (optimal contact) and with the outcome variable (GSJ). Full correlation tables for each study appear in the supplementary material.

Table 2.

Zero-order correlations between key variables (Studies 1a–1e).

Variable	Independent variable: Optimal contact
Variable	Study 1aU.S.	Study 1bIsrael	Study 1cU.K.	Study 1d Spain	Study 1eU.S.
GSJ	.37***	.19*	.19**	.13*	.28***
Perceived fairness of housework	.53***	.55***	.55***	.46**	.55***
Feminist identification	-.20*	-.03	.06	-.06	-.01
Household contribution				-.21**	-.14*
Childcare contribution				-.20*	-.06
	Outcome variable: GSJ
Perceived fairness of housework	.36***	.41***	.24**	.22**	.24***
Feminist identification	-.61***	-.39***	-.41***	-.36***	-.51***
Household contribution				-.03	-.02
Childcare contribution				-.00	-.02

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

As expected, across studies, we found a significant positive correlation between optimal contact and GSJ: the more the relationship was perceived as corresponding to the definition of optimal contact, the greater was women’s tendency to justify the gender social system. Optimal contact was also positively associated with the expected mediator – perceived fairness of housework, and this mediator was positively associated with GSJ.

The zero-order correlations further revealed that women who experienced their relationship as positive also reported doing less housework (particularly in Study 1d, less so in Study 1e). Thus, even though, overall, participants in Studies 1d and 1e experienced inequality within their homes, those with higher levels of optimal contact reported relatively less inequality. In this type of cross-sectional design, this finding could also indicate that optimal contact predicts perceptions of one’s housework division as fair, because it is indeed more fair. In the next studies, we attempted to account for this explanation by manipulating the positivity of contact (Study 2), and the way the housework division is perceived (Study 3).

Next, to examine whether the predicted associations remain significant after taking into account the control variables, we ran a series of partial correlations controlling for age, relationship status, relationship duration, SDO, and household and childcare tasks contributions (the latter variables were controlled for in Study 1d and 1e). We accounted for participants’ reported contribution to the housework, in order to cope with the significant correlations that were found between optimal contact and these two items. The partial correlations between optimal contact and GSJ, controlling for all of the variables above, remained significant and even became stronger, r(1a) = .46, p < .001; r(1b) = .22, p = .012; r(1c) = .23, p = .001; r(1d) = .28, p = .002; r(1e) = .36, p < .001. The partial correlations between optimal contact and perceived fairness of housework, while considering the control variables, remained significant as well, r(1a) = .51; r(1b) = .54; r(1c) = .47; r(1d) = .42; r(1e) = .53, ps < .001.

Mediation analyses

In order to test the predicted indirect effect, we examined whether perceived fairness of housework mediated the association between optimal contact and GSJ by using Hayes’ (2017) PROCESS (model 4). We accounted for the same control variables as in the partial correlations above. Across studies, perceived fairness of housework was significantly associated with GSJ, and the 95% confidence interval of the indirect effect, linking optimal contact to GSJ via perceived fairness of housework was significantly different from zero. This was also the case when the control variables were not added. Table 3 summarizes the results for the indirect effect across studies (see full model in supplementary material online).

Table 3.

Bootstrapped unstandardized indirect effect, SE, and 95% CI of optimal contact on GSJ with Perceived Fairness of Housework as a Mediator (Studies 1a–1e).

Study	Indirect effect	SE	95% CI
1a (U.S.)	0.13	0.04	[0.04, 0.23]
1b (Israel)	0.21	0.05	[0.11, 0.34]
1c (U.K.)	0.07	0.04	[0.00, 0.15]
1d (Spain)	0.09	0.03	[0.03, 0.17]
1e (U.S.)	0.08	0.04	[0.01, 0.16]

Note. Control variables include SDO, relationship duration, relationship status and age. Housework and childcare contributions were accounted for in Studies 1d–1e.

Internal meta-analysis

To examine the robustness of the indirect effect, we conducted an internal meta-analysis. Study 1e was not included in this analysis, due to different sample characteristics (only women currently involved in romantic relationships). We used the unstandardized indirect effects and their standard errors, and conducted the analysis using Comprehensive Meta-Analysis software (CMA version 2). The use of unstandardized indirect effect provides a valid criterion, which is comparable across studies, as our measures were based on the same items-response scales. We performed a random-effects analysis to deal with potential differences in the sample populations. The outcome of the meta-analysis showed that the weighted mean effect (indirect effect of optimal contact on GSJ through perceived fairness of housework) was significant, M = 0.09, 95% CI [0.06, 0.13], Z = 5.68, p < .001.

Moderation analyses

Finally, we tested whether the association between optimal contact and GSJ was moderated by feminist identification (model 1; Hayes, 2017). A moderation analysis (considering all control variables as covariates) revealed a significant interaction effect in two of the samples (Studies 1a and 1b), and a consistent pattern, yet non-significant, in Studies 1d and 1e. Follow up analyses in these four samples indicated that, as expected, optimal contact was significantly correlated with GSJ, among those relatively low on feminist identification. However, this association was weaker (Studies 1a, 1e) or non-significant (Studies 1b, 1d), among those relatively high on feminist identification (see Figure 1 for moderating effect in Study 1b). The interaction between optimal contact and feminist identification in Study 1c (U.K.) did not significantly predict GSJ (see Table 4 for summary of interaction effects; and see supplementary material for the entire model).

Figure 1.

The relationship between optimal contact and GSJ in Study 1b (Israel), as moderated by feminist identification.

Table 4.

Unstandardized Coefficients, SE, t and p value of Optimal Contact and Feminist Identification Interaction on GSJ (Studies 1a–1e).

Study	Coefficient	SE	t	p
1a (U.S.)	-0.08	0.03	-2.41	.017
1b (Israel)	-0.11	0.05	-2.20	.029
1c (U.K.)	-0.01	0.03	-0.50	.612
1d (Spain)	-0.05	0.03	-1.57	.117
1e (U.S.)	-0.04	0.02	-1.72	.086

Note. Control variables include SDO, relationship duration, relationship status and age. Housework and childcare contributions were accounted for in Studies 1d–1e.

Discussion

Correlational evidence obtained from five separate samples across four different countries supported our key prediction. To the extent that women experienced their romantic relationships positively, namely, in a way that corresponds to the definition of optimal contact, they justified gender inequality more.

This association was mediated by participants’ tendency to view the division of housework in their homes as more fair. Results further revealed, in two of the samples, that the more women identified as feminists, the less likley they were to show the predicted association between optimal contact and increased GSJ. Given that feminists are more aware of the existing gender gaps, and the presence of undeniable gender inequality (as indicated by the strong and negative correlations between feminism and GSJ; see also van Breen et al., 2017), we can assume that the experience of optimal contact within their relationships might not suffice to disguise macro-level gender hierarchy.

Studies 1a–1e provide consistent, yet correlative support for our predictions. In order to support the direction of the proposed effect, namely, that women’s tendency to justify gender power relations is derived, in part, from their positive perceptions regarding their own romantic relationships, we conducted two experimental studies. In Study 2, the quality of contact was manipulated, and in Study 3 we directly manipulated the mediator, perceived fairness of housework, in order to validate its causal role in driving the effect (Spencer et al., 2005).

Study 2

In Study 2, we aimed to gain causal support for the “a” path of the suggested mediation model. Thus, the quality of contact was experimentally manipulated, and we tested the effect of this manipulation on women’s perception of fairness regarding their housework division. To avoid ceiling effects, we assumed it would be more feasible to experimentally manipulate a sense of disharmony (rather than a sense of harmony, which Study 1 revealed to be quite high across samples). We therefore included a disharmony condition and a control condition, and predicted that priming negative perceptions of their romantic relationship would lead women to view their housework division as less fair. We further predicted that this decreased perception of fairness would be associated with lower inclination to justify the gender hierarchy.

Method

Participants

Study 2 was conducted among a sample of U.S. MTurk workers. We recruited 397 women currently involved in a heterosexual relationship. Without prior research on this hypothesis to use as a guide for selecting the sample size, we chose to collect 400 participants in total, which we figured would provide, after exclusions, sufficient statistical power (.80) for detecting a small-to-medium effect (f = .15 to .20). We excluded 44 women who did not respond to the manipulation according to the instructions, so that 353 participants remained for analysis (177 in the disharmony condition and 176 in the control condition).

Procedure and measures

The manipulation of optimal contact included a short writing assignment. Women were instructed to describe a situation related to their current romantic relationship. In the disharmony condition, they wrote about a “bad day” with respect to their romantic relationship (e.g., a falling out, a fight, or any negative interaction that was unpleasant and made them feel distant from their partner). In the control condition, they wrote about a typical day involving their partner (i.e., daily routine; see full instructions in supplementary material online). Then, participants completed the questionnaires in the following order: GSJ, α = .82; measures regarding their personal relationship [duration, perceived fairness of housework (α = .95), optimal contact as a manipulation check (α = .88), and relative contribution to housework and childcare]; SDO, α = .81; additional demographics; and feminist identification, r = .88, p < .001. We report additional variables and analyses that were collected for exploratory purposes in the supplementary material.

Results

Means and standard deviations of all study variables, by condition, are presented in Table 5.

Table 5.

Means and standard deviations by conditions for all variables in Study 2.

Variable	Disharmonyn = 177		Controln = 176
Variable	M	SD	M	SD
Optimal contact (manipulation check)	5.46	1.03	5.81	.87
GSJ	4.27	1.11	4.42	1.04
Perceived fairness of housework	4.99	1.57	5.61	1.19
Feminist identification	4.17	1.83	3.68	1.92
Age	38.19	11.59	35.83	11.15
Duration (years)	9.94	9.83	9.10	9.77
SDO	2.52	1.29	2.58	1.28
Household tasks contribution	67.50%	20.38	63.51%	21.29
Childcare contribution	58.94%	31.76	53.14%	32.49%

Note. Statistically significant differences between conditions are marked in bold.

Independent sample t-tests indicated that the groups were not significantly different in terms of relationship duration, t(351) =0.80, p = .42, SDO, t(351) = -0.41, p = .68, or age, t(351) = 1.95, p = .052, even though women in the disharmony condition were slightly older. Importantly, there were no significant differences by condition on the reported amount of household and childcare tasks carried out by the participants, t(351) = 1.69; 1.39, p = .09; p = .16. Regardless of condition, women reported doing more household and childcare tasks than their partners.

An independent samples t-test on the optimal contact manipulation check revealed that, as intended, participants in the disharmony condition rated their relationship as less positive compared to participants in the control condition, t(351) = -3.41, p = .001, d = 0.37. There was also a significant difference between conditions on the perceived fairness of housework measure, t(351) = -4.17, p < .001, d = 0.44, indicating that in the disharmony (vs. control) condition participants saw their housework division as less fair. These differences remained the same, also while accounting for the control variables: age, relationship duration, SDO and household and childcare contributions (see statistics in supplementary material). Thus, the “a” path linking the independent variable (optimal contact) to the mediator (perceived fairness of housework) was causally supported. Notably, women reported doing more of the housework than their partners, regardless of condition. However, when they were primed to think of their romantic relationship as disharmonious, they saw the same division of labor as less fair.

As for GSJ, there was no significant difference between conditions, t(351) = -1.32, p = .18 (this pattern remained while adding the control variables; see statistics in supplementary material). However, when examining the predicted mediation (model 4; Hayes, 2017), we did find the expected significant indirect effect of the manipulation on GSJ, via perceived fairness of housework, B = 0.20, SE = 0.05, 95% CI [0.10, 0.30]. This was also the case when including the control variables: perceived fairness of housework significantly predicted GSJ, and the indirect effect remained significant (statistics are presented in the supplementary material). We could not conduct an analysis to examine the moderation effect of feminist identification, because feminism was affected by the manipulation, t(351) = 2.43, p = .01. Women in the disharmony condition identified more strongly as feminists compared to women in the control condition.

Discussion

Study 2 demonstrated that thinking about a negative incident with their partner (vs. thinking about the relationship in a typical light, which, as Study 1 suggests, is mostly positive), made women less accepting of inequality within their households. This was found even though women in both conditions (disharmony and control) reported doing a similar amount of housework, which in both cases was more than what their partners were doing. Thus, the extent to which women perceive their housework division as fair (or unfair) does not necessarily reflect the actual share of housework they perform. Rather, the way women view the quality of their romantic relationships can shape their fairness perception regarding gender-based inequality at home. This is a fundamental component in our theorizing.

While we did not find an effect of the manipulation on GSJ, Study 2 did replicate the indirect effect linking the manipulation to GSJ via perceived fairness of housework. In addition, even though not expected, women in the disharmony condition were more likely to consider themselves as feminists – an effect consistent with our theorizing reagrding the sedative effects of intergroup contact (Saguy, 2018). Nevertheless, to gain more insight into the causal role of the mediator in driving GSJ, in Study 3 we directly manipulated perceptions regarding housework division as more or less justified.

Study 3

The goal of Study 3 was to test whether women’s perceptions regarding the gender social system are affected by the extent to which they perceive inequality within their homes as fair. We expected that inducing (compared to reducing) women’s tendency to justify their division of housework would enhance their GSJ. We also included an empty control condition, and expected levels of GSJ in this condition to fall in between the two other conditions. The role of feminist identification as a moderator between the condition and GSJ was also examined.

Method

Participants

Study 3 was preregistered (https://aspredicted.org/blind.php?x=gb2fp9) and ran using U.S. MTurk workers.

Based on our sampling plan, we aimed for 100 participants per condition. Given the expected rate of exclusions, we assumed that about 120 participants per condition would provide us with at least 100 participants per condition for analyses. We therefore recruited 361 participants. We excluded 24 participants who did not fill in our main dependent variable measure (GSJ), and three participants who exceeded the predetermined time limit of 40 minutes. After applying these exclusions, we had a sample of 334 participants (110 in the high justification condition; 107 in the low justification condition; 117 in the control condition). All participants answered the written segment according to the instructions, thus, no further exclusions were needed.

Procedure and measures

First, all participants indicated their relative contribution to the housework, compared to their partner. The contribution items were located before the manipulation to avoid potential impact of the manipulation on these reports. Then they were all presented with the following statement: “People have different thoughts and feelings about the division of household chores at their homes. Sometimes the division of work seems logical and sensible, and other times it feels like it should be different.” In the high justification condition, participants were then asked to explain why the current division of labor between them and their partner makes sense in the context of their own lives. Participants in the low justification condition were requested to explain why the current division of labor between them and their partner needs to be restructured (see full instructions in supplementary material). In the control condition, participants moved on to complete the outcome measures. These included the measures of perceived fairness of housework (as a manipulation check; α = .97), GSJ (α =.88), personal details regarding their relationship (including optimal contact, α = .90, and relationship duration), and SDO (α = .83). Participants then filled in their demographic details, in addition to the feminist identification items (r = .88, p < .001). We report additional variables that were collected and analyzed for exploratory purposes in the supplementary material.

Results

Means and standard deviations of each variable are presented, by condition, in Table 6. One-way ANOVAs indicated that the groups did not differ in terms of age, F(2, 331) = 0.44, p = .64; relationship duration, F(2, 330) = 0.24, p = .78 and SDO, F(2, 331) = 0.73, p = .48. In addition, the groups were not significantly different in terms of household contribution, F(2, 329) = 0.23, p = .79, and childcare contribution, F(2, 269) = 0.18, p = .83, with the means indicating that across the three conditions women, again, reported doing on average more housework (M = 4.03, SD = 0.91 on a 1–5 scale) and childcare (M = 3.86, SD = 0.85 on a 1–5 scale) than their male partners.

Table 6.

Means and standard deviations by conditions for all variables in Study 3.

Variable	Low justificationn = 107		Controln = 117		High justificationn = 110
Variable	M	SD	M	SD	M	SD
Perceived fairness of housework (manipulation check)	4.23^a,c	1.82	4.89^b,c	1.66	5.47^a,b	1.59
GSJ	4.00^d	1.17	4.29	1.28	4.41^d	1.22
Optimal contact	5.55	1.00	5.79	.97	5.85	1.02
Feminist identification	3.65	1.90	3.58	1.90	3.71	1.84
Age	38.20	9.07	38.38	9.68	37.23	10.64
Duration (years)	10.89	8.44	10.48	8.33	11.30	9.80
SDO	2.40	1.34	2.43	1.39	2.61	1.22
Household tasks contribution	3.99	0.89	4.02	0.92	4.07	0.93
Childcare contribution	3.90	0.81	3.82	0.92	3.86	0.84

Note. Variables with significant differences between conditions are marked in bold. Significantly different means are denoted by superscripted matching letters.

The manipulation of justification of housework division was successful, F(2, 331) = 14.13, p < .001, η² = .08. Tukey follow-up comparisons revealed that participants in the high justification condition rated the division of housework as more legitimate, compared with participants in the control (p = .03, d = 0.35) and low justification conditions (p < .001, d = 0.71). In addition, participants in the control condition rated the division of housework as more legitimate than participants in the low justification condition (p = .01; d = 0.37).

As expected, the manipulation affected participants’ GSJ, F(2, 331) = 3.09, p = .046, η² = .02. Tukey follow-up comparisons revealed that the level of GSJ among participants in the high justification condition was significantly higher, compared to the low justification condition (p = .042, d = 0.34). Levels of GSJ in the control condition fell between the two experimental conditions, but were not significantly different from them. Results remained significant while adding the control variables (see statistics in supplementary material).

We also examined the role of feminist identification as a possible moderator between condition (high/low justification) and GSJ (model 1; Hayes, 2017). Results indicated a significant main effect for feminist identification (B = -0.32, SE = 0.08, t = -3.83, p < .001), but not for the condition (B = 0.27, SE = 0.15, t = 1.79, p = .07). The interaction effect was non-significant (B = -0.02, SE = 0.04, t = -0.44, p = .66), and remained non-significant when we added the covariates to the model (see supplementary material).

Discussion

The findings of Study 3 suggest that increasing women’s tendency to legitimize (vs. challenge) the division of housework between them and their romantic partner, irrespective of whether the distribution is equal or not, drives them to view gender relations in overall society as fair. This further validates our assumed generalizing effect from the personal romantic relationship to the entire groups involved, that is, men and women in general.

General Discussion

Women are disadvantaged relative to men across critical life domains pertaining to their basic rights, earning potential, and opportunities for social advancement. Yet, relative to other disadvantaged groups, women have a low awareness of group-based discrimination, and they readily justify gender hierarchy (Harnois, 2015; Jost & Kay, 2005). In the current research, we tested the hypothesis that romantic relationships may be one of the roots of this tendency.

In five correlational studies, conducted in Western countries, we showed that to the extent that women hold positive views about their own romantic relationship, they tend to view the (mostly unequal) housework distribution between them and their male partner as more fair. This fairness perception within the relationship itself, then, predicted a greater inclination to justify the gender hierarchy in general.

Our argument included two components. The first linked positive views regarding one’s own relationship to perceptions regarding inequality within one’s household. The second linked perceptions regarding inequality within one’s household to perceptions regarding gender hierarchy in general. Experiment 2 provided causal support for the first component by showing that a manipulation of the quality of the romantic relationship shaped perceptions of fairness regarding housework division. Notably, in both conditions, participants reported a similar amount of housework. Thus, the effect of the condition on perceptions of fairness cannot be attributed to differences in actual housework contribution. Experiment 3 provided causal support for the second component, showing that a manipulation of fairness perceptions regarding one’s housework, directly affected gender system justification.

We further expected the predicted effects to be less pronounced among women who identify as feminists. However, feminist identification moderated the association between optimal contact and gender system justification in only two of the correlational studies. Given that feminism might be understood differently by different people in different places, and that some women may be reluctant to own the title “feminist” (Duncan, 2010), the moderating role of feminist ideology could be better assessed in future research. This can be done by using a more nuanced measure that more accurately taps into women’s discontent with the present state of gender inequality. Another possible explanation for why feminism did not moderate the effects consistently is that women’s feminist attitudes might, ironically, lead them to be equally likely to rationalize household inequality. Given the inconsistency between feminist ideals and house-related inequality, justifications for such inequality might be readily used among feminist women (see Yurtsever et al., 2021). If this is the case, we would not expect the manipulation used in Study 3 to affect feminist women and less feminist women differently.

Link to Previous Theories

Together, our findings reinforce the irony of harmony effect (Hässler et al., 2020; Saguy, 2018), according to which optimal contact can lead disadvantaged group members’ to have less support for social change. The present research expands the scope of the irony of harmony effect beyond ethnic and racial relations to the gender domain, and specifically to romantic relationships. This expansion highlights two insights: romantic relations can be seen as an intergroup encounter; and the romantic relationship itself can, quite paradoxically, nourish social hierarchy.

Previous work on women’s tendency to justify the gender hierarchy showed that benevolent sexism increases gender system justification (Jost & Kay, 2005), and decreases women’s willingness to partake in collective action (Becker & Wright, 2011). Our findings are in line with these effects, because benevolently sexist views and related behaviors are likely to emerge within romantic relationships, and contribute to the stability of gender hierarchy.

More broadly, the current findings echo feminist arguments regarding the impact that romantic relationships can have on the development of a collective consciousness among women (e.g., de Beauvoir, 1953; Hacker, 1951; Rossi, 1969). Feminist scholars have emphasized the difficulty for women to band together as long as they are positively attached to advantaged group members. Jackman (1994) added that within romantic relationships, advantaged group members can cover inequality using warmth and personal affection. According to this approach, the intimate closeness achieved through romantic relationships could function, in part, as one source behind the stability of gender power relations. This is much in line with our findgins, showing that the harmony men and women share within romantic relationships can lead women to be more accepting of gender inequality. We further pointed to a possible mechanism for this effect: women’s tendency to generalize fairness perceptions from their own relationships to the outside world. Similar generalization processes were documented in the abundance of work on intergroup contact (see Pettigrew, 1998, for a review). However, as far as we know, the present research is the first to apply this notion to interpersonal romantic contact between men and women, and first to give empirical evidence to these ideas.

Implications, Limitations, and Future Directions

Our findgins raise the challange of how to undermine women’s tendecy to legitimize gender inequality. Although raising awareness to inequality within one’s home can constitute a potential solution (as suggested in Study 3), this could also possibly harm the romantic relationship. Paths for interventions should probably set aside practices pertaining to the relationship, and instead consider how to raise awareness of gender inequality more generally, in outlets such as educational settings (e.g., gender education, dialogue groups or lectures), and how to involve men as well, as potential agents of change (Becker et al., 2013; Wright & Lubensky, 2009).

The reported studies have several limitations to consider. First, the samples used are non-representative and involve mostly Western, Educated, Industrialized, Rich and Democratic (WEIRD) women (Henrich et al., 2010). Thus, the ability to generalize our findings to other populations is limited. Second, our use of self-report measures is not ideal, given limitations associated with consistency motives, self enhancement, and self-presentation considerations (Swann et al., 2007). Future work can benefit from using more implicit measures for assessing women’s perceptions regarding gender relations, or by combining behavioral measures.

Even though we included two experiments, the bulk of our evidence is correlational, thus, our ability to infer causality is limited. In addition, in Study 2, we did not find a main effect for the manipulation of contact’s positivity on GSJ. It seems that a single manipulation was not sufficient to directly affect women’s general perceptions regarding the gender social system. We assume that the effect of optimal contact on GSJ is formed gradually, while the romantic relationship develops. Given these limitations, it could be useful to conduct a longitudinal study, in which women’s perceptions will be assessed over time and in different stages of the relationship. It can also be helpful to ecologically validate the effects by exploring them in settings that are closer to real-life interactions, such as couples’ studies.

Conclusions

Many individuals desire a romantic relationship involving common goals, equality, and a feeling that our partner knows us deeply. Our research shows that this type of desired relationship can result in women viewing their own disadvantaged position (at home, and in society) as acceptable. This view can result in less willingness to act for equality, and therefore potentially contribute to the preservation of hierarchical relations between men and women. Taken together, this notion helps reveal how gender inequality is maintained in ways that involve both men and women, and points to the importance of raising awareness of various domains of inequality (within and outside the home), and how they might be interconnected.

Supplemental Material

sj-docx-1-gpi-10.1177_13684302221100403 – Supplemental material for The irony of (romantic) harmony: Heterosexual romantic relationships can drive women’s justification of the gender hierarchy

Supplemental material, sj-docx-1-gpi-10.1177_13684302221100403 for The irony of (romantic) harmony: Heterosexual romantic relationships can drive women’s justification of the gender hierarchy by Danit Sobol-Sarag, Noa Schori-Eyal, Saulo Fernández and Tamar Saguy in Group Processes & Intergroup Relations

Footnotes

Funding

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

ORCID iDs

Noa Schori-Eyal

Tamar Saguy

Supplemental material

Supplemental material for this article is available online.

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