Baseline Results

Table 2 reports the intergenerational transmission coefficients for the sample of ownbirth (upper panel) and adoptive (lower panel) children. We use an indicator for being nominated at least once as our main outcome.Footnote 10 We use linear probability models to facilitate comparisons with earlier adoption studies. However, as shown in the Online Appendix, the findings are substantively identical if we instead use a logit model. We report separate models for whether either parent (column 1), the father (columns 2 and 4), or the mother (columns 3 and 5) ran for office. In the sixth column we enter both paternal and maternal candidacy.

Throughout, we report estimates from models that include a set of baseline covariates. These covariates include the child’s gender, birth-year indicators for the child and each parent, and 24 county indicators of where each parent lived at the time of the child’s birth. Apart from these individual level controls, we also include the ratio of local council seats to the electorate in the municipality within which the child resides at the time of the election (Seats-to-Voters).Footnote 11 As already noted, the variation in our outcome measures is mainly driven by candidates nominated at the municipality level. Moreover, local Swedish elections operate by a party-list system where local nomination committees largely control who gets nominated and how candidates are ranked on the list. Given that the number of candidates included on these party lists far from perfectly reflects the number of voters in a municipality, there will be a strong and mechanically negative relationship between the size of the electorate in the municipality and the chance of being nominated (Besley et al. 2013; Dancygier et al. 2015).

Consider, first, the intergenerational transmission of political candidacy in the ownbirth sample. The estimates in column 1 suggest that having a parent who ran for office is associated with an approximately 5 percentage points higher probability of standing as a candidate. Given a baseline probability of 2.27% among the ownbirth children (see Table 1), the magnitude of this transmission should be considered very large. According to columns 2 and 3, the candidate experience of mothers is somewhat more important. The likelihood of running for office is more than 6 percentage points higher for those having a politically active mother. The corresponding estimate associated with having a father run for office is 5.3 percentage points.

To further put the magnitude of the intergenerational transfer in perspective, Table 2 reports the effect of increasing the Seats-to-Voters ratio from the 10th to the 90th percentile. This corresponds to increasing the number of seats in a municipal assembly from one in every 1573 voters to one in every 209 voters. As expected, such a shift has a large impact on the probability of running for office. However, the size of this mechanical effect (2.4 percentage points) is less than half the size of the intergenerational transmission rate. Having politically active parents is thus a very important predictor of offspring candidacy status.

In columns 4–6 we turn to the potential bias stemming from parental concordance. Recent research has shown that the interspousal concordance in political attitudes is among the strongest of all social and biometric traits, including educational attainment, height and church attendance (Alford et al. 2011). Consistent with these findings, the tetrachoric correlation between mother’s and father’s candidacy status in the ownbirth sample is strong: r = 0.50,\(p<0.01\).

The entries in column 6 display estimates from a model that includes both fathers’ and mothers’ candidacy experience simultaneously. For reasons of comparison, columns 4 and 5 report results from separate paternal and maternal models with the sample restriction that both mothers and fathers are non-missing. Reflecting the high degree of spousal concordance, the transmission coefficients from the separate paternal and maternal models reported in columns 4 and 5 (in the upper panel of Table 2) decrease in magnitude when including both parents’ candidacy status as covariates in the same model (column 6). The paternal effect goes from 0.053 to 0.045 and the maternal effect falls from 0.063 to 0.053. Thus, having both a mother and a father who ran for office is associated with an almost 10 percentage points higher probability of standing as a candidate.

The lower panel of Table 2 reports the results for the adoption sample. The estimates show that both pre-birth and post-birth factors are significant predictors of the adoptee’s likelihood of standing as a candidate. According to column 1, the pre-birth and post-birth effects are approximately equal in size. Holding the rearing parents’ candidacy status constant, the probability of running for office is 3.5 percentage points greater if one or both of the biological parents were nominated (\(p<0.01\)). Holding constant the biological parents’ candidacy experience, a child who had at least one rearing parent who ran for office is 3.4 percentage points more likely to stand as a candidate (\(p<0.01\)).

The estimates of father–child transmission in column 2 suggest that the pre-birth effect is almost twice as large as the post-birth effect. Strikingly, however, the mother–child association seems to be mainly due to a strong post-birth effect combined with a substantially weaker and statistically insignificant impact of the biological mother’s candidacy status. The sum of the biological and adoptive parent transmission coefficients is presented at the bottom of Table 2. Across the three specifications presented in columns 1, 2, and 3, the sum of the pre-birth and post-birth effects is slightly larger than the corresponding intergenerational association for ownbirth children in the upper panel. In no case, however, is the difference statistically significant.

Turning again to parental concordance, the interspousal correlations for candidacy status in the adoption sample are equal to 0.18 (biological mother–biological father, \(p=0.012\)) and 0.53 (adoptive mother–adoptive father, \(p<0.01\)). The fact that the correlation for the adoptive parents is much stronger than the corresponding correlation for the biological parents suggests that the upward bias in the paternal and maternal transmission coefficients in the upper panel in Table 2 is mainly driven by inflated post-birth effects. The results presented in columns 4–6 in the lower panel of Table 2, based on a sample restricted to children for whom we have information on both biological and adoptive mothers’ and fathers’ candidacy status, confirm this conjecture. Since many biological fathers are missing, the sample size declines significantly in comparison to the samples used in the separate paternal and maternal models (shown in columns 2 and 3). When taking spousal concordance in candidacy status into account (column 6), the post-birth effects decrease in size whereas the pre-birth effects are less affected.

Several important lessons can be drawn from these results. First, there is a strong parent–child transmission in the tendency to run for office. The baseline probability of standing as a candidate of 2.27% and the estimates of the transmission coefficient imply that having at least one parent who is or was a candidate is associated with a doubling of the likelihood of running for office.

Second, when considering the adopted children, we find evidence that the intergenerational transmission in candidacy status is composed of both pre-birth and post-birth factors. Having a biological parent who ran for office is a good predictor of the adoptee’s probability of running for office. This is despite the fact that all formal links between biological parents and children were broken at the time of adoption and a large majority of the children in the sample have no information about their biological parents (Nordlöf 2001).

Third, the estimates also indicate that adoptive parents’ political activity is a major source of intergenerational resemblance. All specifications show positive and significant transmission estimates for rearing fathers and mothers. Judging by the results based on the joint parental indicator in column 1 or the combined estimates in column 6 (of Table 2), approximately half of the intergenerational association in candidacy status is accounted for by pre-birth factors and half by post-birth factors.Footnote 12

Fourth, the transmission from mothers to children seem to be somewhat stronger than the father–child association. Furthermore, whereas the paternal effect is driven by both pre-birth and post-birth factors, the link between mothers and their children is almost fully accounted for by the post-birth environment. Below, we argue that this pattern of results is consistent with a role modeling mechanism.

Finally, in order to avoid inflated transmission estimates due to the strong interspousal correlation in the tendency to run for office, it is important to include both mothers’ and fathers’ candidacy status jointly in the models.

In the Online Appendix we discuss and provide evidence of the internal and external validity of these results. In terms of internal validity, we show that (i) the small fraction of the adoptees that were placed with their adoptive parents after turning one year old are unlikely to bias the pre-birth and post-birth estimates to any great extent; (ii) the transmission estimates are not sensitive to non-random placement based on a large set of observable parental characteristics; (iii) restricting the sample to adoptees whose biological and adoptive parents lived in different counties does not alter the results; (iv) the transmission estimates are robust to using nominated but not elected parents; and v) reweighting the adoption sample such that the birth and adoptive parents are more similar to the parents in the ownbirth sample only marginally influences the transmission coefficients. As for external validity, we present transmission estimates for a number of political traits related to candidacy status in a Swedish and a US sample. These estimates are strikingly similar across the two countries.

Table 2 Transmission coefficients for political candidacy Full size table

Pre- and Post-birth Mechanisms

Thus far we have presented results demonstrating that both pre-birth and post-birth factors predict candidacy status and that each account for approximately equally sized shares of the intergenerational association in the tendency to run for office. In the next step of the analysis, we investigate possible mechanisms that may mediate the pre- and post-birth effects. Turning first to the pre-birth effects, a common hypothesis is that any genetic effects on complex behavioral traits will be indirect and mediated by different psychological mechanisms such as cognitive ability and personality traits (Mondak et al. 2010). This hypothesis has been supported by previous research on mass political participation (Dawes et al. 2014, 2015).

To test whether cognitive ability and personality also mediate the pre-birth effects on elite political participation, we use two measures based on conscription data provided by the Military Archives of Sweden.Footnote 13 Our measure of cognitive ability is based on the results from four subtests intended to capture logical, verbal, spatial, and technical abilities. To construct the indicator, we summed the scores of the four subtests and standardized the index such that the mean is equal to zero and the standard deviation is equal to one. The resulting measure has been shown to be a good measure of general intelligence (Carlstedt 2000) and positively predicts candidacy status in the Swedish context (Dal Bó et al. 2016).

Our measure of personality is based on the results from interviews with the conscripts conducted by psychologists. The main objective of the interview was to certify that the conscript could cope with the psychological requirements of the military service and, ultimately, war stress. Based on the interview, the psychologist ranked the conscript’s military aptitude along a nine-point Stanine scale with a mean of five and a standard deviation of two (Lindqvist and Vestman 2011). The measure we use in the analysis is a standardized version of this scale with mean zero and unit variance. It is important to note that this is a measure of a specific ability (military aptitude) rather than a specific trait. Military aptitude has been shown to tap into a bundle of different personality traits that we should expect to be positively related to candidacy status such as willingness to assume responsibility, emotional stability, independence, persistence, having an outgoing character and power of initiative (Fox and Lawless 2005; Lawless 2011; Kanthak and Woon 2015; Schneider et al. 2016). In addition, the measure of military aptitude used in this study has been reported to be highly correlated with leadership skills (Lindqvist and Vestman 2011).

We conduct a very simple test of the mediation hypothesis by comparing the estimated pre-birth effects when controlling and not controlling for cognitive ability and military aptitude. A significant decrease in the magnitude of the pre-birth effects would suggest mediation. Since conscription was mandatory only for males, the analysis is restricted to the male children in both samples. The results are presented in Table 3. To save space, we focus on models using the joint parental indicators.

As a benchmark, column 1 displays the transmission estimates from a model restricted to male children for whom we have information on cognitive ability and military aptitude. First, we can see that the transmission estimates from this restricted sample are very similar to the ones presented in column 1 of Table 2. Above all, the strong intergenerational transmission in candidacy status reflects both pre-birth and post-birth factors. Second, the results in column 2 suggest that cognitive ability is strongly related to candidacy status. A one standard deviation increase in cognitive ability is associated with a 0.6 percentage point increase in the probability of running for office in the ownbirth sample. The corresponding effect in the adoption sample is twice as large (0.013). Moreover, the coefficients for cognitive ability are weakly statistically different across the ownbirth and adoption samples (\(p=0.09\)).Footnote 14 The effect of military aptitude is weaker and less precisely estimated, especially in the adoption sample. More importantly, controlling for cognitive ability and military aptitude does not alter the magnitude of the pre-birth effect. The estimated effect of having at least one parent who ran for office is 0.051 in both models. Consequently, the pre-birth transmission in candidacy status does not appear to be mediated by cognitive and non-cognitive skills as measured by the conscription tests.

Turning instead to the post-birth mechanisms, an important question concerns the extent to which the estimated post-birth effects reflect a causal link between political activity of the rearing parent and child. Such a pathway would operate whenever parents serve as role models that children emulate when creating their own political identities. Another possible mechanism explaining the link between parent and child political behavior is if parent activity levels shape the participation-relevant experiences and skills that their children acquire (Westholm 1999).

We can only indirectly and partly test these assertions with the data at our disposal. In a first step, we attempt to rule out the possibility that the parent–child link in candidacy status is driven by intergenerational associations in related attributes or traits. The most likely confounder in this instance is educational attainment. Based on previous research, we know that there is a significant association between education and the likelihood of running for office (Dancygier et al. 2015). Past studies have also reported a strong link between parental and child educational attainment (Björklund et al. 2006). Consequently, the estimated post-birth effects presented in Table 2 may reflect intergenerational transmission in education rather than a direct causal pathway between parent and child political activity.Footnote 15 In a second step, we estimate separate models for sons and daughters in order to explore the extent to which the post-birth effects on candidate status reflect a role modeling mechanism.

The results of the first step are presented in columns 3 through 5 in Table 3. Column 5 reports estimates from a model controlling for parent and child years of schooling.Footnote 16 The years of schooling variable has been recoded to the 0–1 range such that 0 denotes the sample minimum (7) and 1 the sample maximum (19). Columns 3 displays results from a corresponding baseline model with the sample restrictions implied by the inclusion of parental and child education, whereas column 4 presents estimates from a model including parental but not child education.

In the upper panel we can see that parental education is significantly related to the probability of standing as a candidate (model 4) and that this effect is almost entirely mediated by one’s own education (model 5). The pattern of results is similar in the adoption sample, although less precisely estimated. Comparing across models 3 and 5 it is also clear that the estimated transmission coefficients change only marginally, if at all, when controlling for parent and child years of schooling. Thus, the intergenerational transmission in candidacy status does not seem to be driven by the intergenerational association in educational attainment despite the fact that education itself is strongly related to candidacy status.

Next, we turn to the argument that the post-birth effects we observe may partly be explained by a role modeling mechanism. Recent research has shown that exposure to female role models promotes women’s political interest, ambition, and engagement. Atkeson (2003) reports an increased level of political engagement among women living in U.S. states with visible and competitive female candidates, whereas no such effect is detected among men in states with competitive all male races. Similarly, in two cross-national studies, Campbell and Wolbrecht (2006) and Wolbrecht and Campbell (2007) find that the presence of viable female political candidates and female members of parliament positively influences adolescent girls’ anticipated and adult women’s actual political involvement. Using Swiss data, Gilardi (2015) finds that the election of a woman in a given municipality increases the probability of female candidates in neighboring municipalities in the next election. On the other hand, Broockman (2014) fails to find any effect of a female candidate’s winning an election on other women running for office in U.S. state legislative elections.

Against this backdrop, we expect stronger transmission between mothers and daughters than between either mothers and sons or fathers and any child. Furthermore, the relative magnitudes of post-birth transmission rates across the different parent–child constellations, but not necessarily the pre-birth effects, should correspond to the pattern of transmission coefficients found in the ownbirth sample. To test this hypothesis, we allow the effects of politically engaged mothers and fathers to vary between daughters and sons.

Columns 6 and 7 in Table 3 display transmission estimates from such models. The results lend some support to the expectations. Above all, the post-birth transmission in candidacy status seem mostly to be driven by the influence of rearing mothers on their daughters. Looking first at the transmission estimates based on the ownbirth sample, the probability of standing as a candidate is 5.8 percentage points higher for daughters whose mothers ran for office. This effect is significantly stronger (\(p<0.01\)) than any of the other three transmission coefficients. Although less precisely estimated, the pattern of post-birth effects displayed in the lower panel is similar to the one obtained in the much larger ownbirth sample. The transmission from rearing parents to daughters appears to be predominantly due to the maternal influence (\(p=0.051\)). The adoptive mother–daughter effect is larger in magnitude than any of the other post-birth effects. However, in no case are these differences statistically significant. Finally, as was the case in the baseline models, the estimated pre-birth effects are clearly related to the sex of the parent. The influence of having a birth father who ran for office is positive whereas the maternal pre-birth effects are small in magnitude and insignificant. It should be noted, though, that the paternal pre-birth effects are somewhat stronger than (although not statistically different from) the corresponding maternal pre-birth effects.