Abstract
Many studies during the past 50 years have found an association between father absence and earlier menarche. In connection with these findings, several evolutionary theories assume that father absence is a causal factor accelerating reproductive development. However, a recent study analysing data from the Avon Longitudinal Study of Parents and Children (ALSPAC) found that father absence does not predict age at menarche when adjusted for sibling relatedness. In this study, we have replicated these results in the Czech section of the European Longitudinal Study of Pregnancy and Childhood (ELSPAC), which used the same questionnaires as ALSPAC to study a geographically distinct population. Our results support the conclusion that sibling relatedness rather than father absence predicts age at menarche. Furthermore, our results show that age at menarche in 1990s UK and Czech cohorts is very similar despite socioeconomic differences between the two countries.
Keywords: menarche, sibling relatedness, father absence, inclusive fitness, ELSPAC
1. Introduction
Menarche is an important point in female development with far-reaching influence on the rest of life. Earlier menarche is associated with increased all-cause mortality [1], a risk of breast cancer [2], higher BMI in adult life [3], metabolic syndrome [4] and other outcomes [5]. This relationship is not as surprising as it may seem since classical evolutionary theories of ageing postulate that ageing begins once sexual maturity is reached [6,7]. Thus, it may be argued that earlier menarche translates almost directly to a ‘faster’ pace of ageing, which could explain all associated phenotypes.
Several factors such as weight [8], ethnicity [9], socioeconomic status [10] and others [11] are known to modify age at menarche. One factor that is especially intriguing from an evolutionary biology perspective is father absence [12–15]. Several evolutionary explanations of its association with earlier menarche have been proposed [12,13,16,17]. Most of them assume that the timing of menarche is positively associated with the timing of fertility. Some of them suggest that father absence is a signal of a future environment where it is beneficial to reproduce early [16,17], while others suggest that father absence reduces parental investment, causing offspring to invest in reproduction rather than in continued growth [12,13]. However, the entire notion that father absence leads to earlier menarche has recently been disputed. A 2017 paper by Daniel Smith analyses data from the Avon Longitudinal Study of Parents and Children (ALSPAC) [18] and shows that father absence does not lead to earlier menarche when accounting for biological relatedness between siblings [19]. Smith interpreted these results in the framework of the inclusive fitness theory [20], arguing that an individual is more likely to postpone reproduction and invest in siblings if these are full siblings (r = 0.5) because the indirect fitness benefits are much greater than investing in half-siblings or step-siblings (r = 0.25 or 0 respectively) [21]. While the association between sibling relatedness and menarche described by Smith is robust, it only focuses on a single Western European cohort and stands in direct opposition to many previous studies. Furthermore, the age of menarche is affected by a complex web of socioeconomic factors [22–24]. Therefore, testing whether association reported by the Smith is also present across geographically distinct populations living in different socioeconomic conditions is of significant interest. The Czech cohort included in the European Longitudinal Study of Pregnancy and Childhood (ELSPAC) [25], which used questionnaires identical to those employed by ALSPAC, constitutes ideal material for an attempt at replicating Smith's work. This study therefore tested the relationship between sibling relatedness and menarche in the Czech ELSPAC cohort.
2. Material and methods
(a). Study subjects
Data were obtained from the Czech arm of ELSPAC, initiated by the World Health Organization Regional Office for Europe [25]. The original project included eight independent centres based in the present-day Czech Republic, Greece, Russia, Slovakia, Ukraine and the United Kingdom (UK). Unfortunately, currently, every national ‘arm’ of ELSAPC manages its own data, making it almost impossible to get pooled data from each country involved. For the Czech section of the study, all mothers were recruited in the Brno and Znojmo regions of the present-day Czech Republic and were expected to deliver between 1 March 1991 and 30 June 1992. The cohort included 3548 families with female children [26].
Age at menarche was assessed using a series of questionnaires completed at ages 11, 13, 15, 18 and 19 during the course of general practitioner appointments and confirmed by questionnaires filled in by mothers at ages 15, 18 and 19. While questionnaires at ages 11, 13 and 15 contained questions about precise age at menarche, questionnaires at 18 and 19 were limited to establishing whether menarche had already occurred. Therefore, if menarche was listed as not having occurred at age 15 but as having occurred already at 18, we excluded the girl from the analysis. This happened in the case of 18 girls out of the total of 1151.
The age of menarche has been determined with a resolution of to the nearest month. Father absence status was assessed according to a questionnaire filled out at the age of 7 based on the girls' mothers. The questions used to determine father absence were: What is your current marital status? Is your current partner the biological father of the child? Do you currently have a partner? If you currently have a partner do you live together? A total of 224 girls in our sample had an absent father. The number of siblings was assessed at the girl's age of 15.
The final analysis is based on data for 1151 girls with known ages at menarche and fewer than four missing variables. These variables were: number of siblings, sibling relatedness, father absence, birth weight, maternal age at menarche, maternal education prior to delivery, and equivalized household income, which is a widely used measure adjusting for the number of people in each household [27].
(b). Statistics
Descriptive statistics (electronic supplementary material, table S1) of continuous variables are expressed as mean ± standard deviation (s.d.). In the case of categorical variables, absolute and relative frequencies are used. The normality of continuous variables was checked using statistical tests (e.g. Shapiro–Wilk, Pearson, Anderson–Darling) as well as graphical tools (Q–Q plot, histogram). In the case of severe violations of normality, the variable in question was logarithmically transformed and normality was checked again. Equivalized family income was the only log-transformed variable. For the purpose of intergroup comparisons (for continuous variables) only parametric tests were used (t-test, ANOVA) (electronic supplementary material, table S2). Univariate (one independent variable) and multivariable/multiple (more independent variables) linear regression models with age at menarche as the dependent variable were constructed. β represents the regression coefficient measured in years. Missing values were imputed using Multivariate Imputation by Chained Equations (MICE). Variables used in the imputation procedure were age at menarche, siblings relatedness, father absence, birth weight, maternal education, equivalized family income, number of siblings and maternal age at menarche. 18.1% of the data were imputed. Missing data were not missing completely at random (MCAR) (Little's test: X2 = 206.1, d.f. = 127, p < 0.001). All models used the same imputed datasets. All analyses were performed using statistical software R, version 3.5.1 (https://www.r-project.org/). We have considered p-values less than 0.05 as statistically significant.
3. Results
Average age at menarche (figure 1) in our sample stood at 12.77 (s.d. = 1.00) years. Average age at menarche for participants with full siblings only was 12.82 (s.d. = 1.00), while for those with only half/step-siblings or with no siblings, it was 12.48 (s.d. = 0.90) and 12.53 (s.d. = 1.02) respectively. Age at menarche in our cohort is similar to the age at menarche in the ALSPAC [19] cohort (table 1).
Figure 1.
Mean age at menarche for every category of sibling relatedness. Error bars denote 95% CI.
Table 1.
The table compares average age at menarche between ALSPAC and Czech ELSPAC cohorts. Data about age at menarche for the ALSPAC cohort were obtained from Smith [19].
| ELSPAC | ALSPAC | |
|---|---|---|
| age at menarche (s.d.) [n] | ||
| full cohort | 12.77 (1.00) [759] | 12.62 (1.17) [2922] |
| siblings | ||
| only full siblings | 12.82 (1.00) [540] | 12.70 (1.10) [2492] |
| only half/step-siblings | 12.48 (0.90) [67] | 12.28 (1.33) [85] |
| no siblings | 12.53 (1.02) [67] | 12.55 (1.14) [187] |
| half/step- and full siblings | 12.87 (0.99) [85] | 12.57 (1.24) [158] |
In a univariate model where study subjects with full siblings are used as a reference group (table 2, model 1), groups of subjects with only half/step-siblings and no siblings exhibited a significantly earlier age at menarche (β = −0.34; 95% CI (−0.56; −0.12); p = 0.002 and β = −0.29; 95% CI (−0.50; −0.07); p = 0.008 respectively). Contrary to much previously published data, we found no significant association between father absence and age at menarche when using the univariate model containing only ‘father absence’ as a criterion (β = −0.03; 95% CI (−0.18; −0.11); p = 0.674). In the third model, which incorporated both ‘sibling relatedness’ and ‘father absence’ (table 2, model 3), groups with only half/step-siblings and no siblings were found to have significantly earlier menarche (β = −0.34; 95% CI (−0.56; −0.12); p = 0.003 and β = −0.29; 95% CI (−0.50; −0.07); p = 0.009, respectively) but no independent effect of father absence was established (β = 0.00; 95% CI (−0.15; −0.15); p = 0.988). These trends remained the same even after including additional variables such as maternal education (table 2, model 4), birth weight, number of siblings, maternal age at menarche and equivalized family income (table 2, model 5; electronic supplementary material, table S3). Models 4 and 5 were separated to mirror the models used for the UK ALSPAC cohort [19]. All models were also recalculated on a dataset without imputations of father absence and sibling structure (electronic supplementary material, table S4); the results of this analysis were very similar to the above-presented results, and using them instead of the imputed dataset leads generally to the same conclusions.
Table 2.
Models investigating the association between sibling relatedness and age at menarche. Models 1 and 2 are univariate while all other models are multivariable. Model 1 includes sibling relatedness and age at menarche. Model 2 includes father absence and age at menarche. Model 3 combines models 1 and 2, i.e. includes both sibling relatedness and father absence. Model 4 includes additional variables: maternal education prior to delivery and maternal age at menarche. In model 5, the list of additional variables is further expanded and includes maternal education prior to delivery, maternal age at menarche, birth weight, number of siblings and equivalized household income. Sample size in all models was imputed to be 1400. Italic type indicates p-value <0.05.
| β-estimate | 95% CI for β est. | p-value | |
|---|---|---|---|
| model 1 | |||
| siblings | |||
| only full siblings | (reference group) | ||
| only half/step-siblings | −0.34 | (−0.56; −0.12) | 0.002 |
| no siblings | −0.29 | (−0.50; −0.07) | 0.008 |
| half/step- and full siblings | 0.04 | (−0.16; 0.24) | 0.699 |
| model 2 | |||
| father absence | |||
| no | (reference group) | ||
| yes | −0.03 | (−0.18; 0.11) | 0.674 |
| model 3 | |||
| siblings | |||
| only full siblings | (reference group) | ||
| only half/step-siblings | −0.34 | (−0.56; −0.12) | 0.003 |
| no siblings | −0.29 | (−0.50; −0.07) | 0.009 |
| half/step- and full siblings | 0.04 | (−0.17; 0.25) | 0.706 |
| father absence | |||
| no | (reference group) | ||
| yes | 0.00 | (−0.15; 0.15) | 0.988 |
| model 4 | |||
| siblings | |||
| only full siblings | (reference group) | ||
| only half/step-siblings | −0.29 | (−0.51; −0.08) | 0.008 |
| no siblings | −0.28 | (−0.49; −0.07) | 0.008 |
| half/step- and full siblings | 0.04 | (−0.16; 0.24) | 0.695 |
| father absence | |||
| no | (reference group) | ||
| yes | 0.00 | (−0.14; 0.15) | 0.959 |
| other variables | yes | ||
| model 5 | |||
| siblings | |||
| only full siblings | (reference group) | ||
| only half/step-siblings | −0.28 | (−0.50; −0.07) | 0.011 |
| no siblings | −0.29 | (−0.54; −0.04) | 0.022 |
| half/step- and full siblings | 0.05 | (−0.16; 0.27) | 0.638 |
| father absence | |||
| no | (reference group) | ||
| yes | 0.00 | (−0.14; 0.15) | 0.956 |
| other variables | yes | ||
4. Discussion
Our findings show a strong association between sibling relatedness and age at menarche. As these findings are in accordance with results previously achieved using the UK ALSPAC cohort [19], our results thus show that this association exists in geographically distinct populations. The main weakness of our replication is a relatively limited number of subjects in the comparator groups, i.e. no siblings, only half/step-siblings, and half/step-siblings and full siblings, and the fact that the study design did not allow us to time precise arrival of siblings. Nevertheless, we believe the cohort size is not excessively small and is sufficient for our aim, which was an independent replication of findings made on a larger cohort. Another weakness of our study is that we were not able to precisely time order and timing of the birth of individual siblings. The main strength of our replication design is that the Czech ELSPAC cohort was recruited simultaneously with the UK ALSPAC cohort and used Czech translations of ALSPAC questionnaires. The use of the Czech ELSPAC cohort in fact enabled us to replicate Smith's results in an identically constructed cohort.
Arguably, the most interesting finding of our study is that the average age at menarche was nearly identical for both cohorts (ALSPAC: 12.62, Czech part of ELSPAC: 12.77). This similarity is quite intriguing in view of the fact that the socioeconomic situation in the UK and the Czech Republic during the 1990s and early 2000s differed significantly. From the outset, one of the aims of the Czech ELSPAC cohort was to investigate the effects of profound socioeconomic changes associated with the societal transformation after the fall of Communism in 1989 [26]. However, no such changes were taking place in the UK at the time; in any case, these profound socioeconomic shifts—or lack thereof—did not influence age at menarche in any way. Furthermore, in the 1990s, there were also significant differences in life expectancy, lifestyle, diet, psychosocial stress and other factors between the UK and the Czech Republic [28]. Our results thus suggest that these differences in socioeconomic factors between the UK and the Czech Republic in the 1990s were not strong enough to significantly alter the age at menarche.
One possible evolutionary explanation of the association between sibling relatedness and age at menarche may be that children are more likely to invest in their mothers' reproduction (and delay they own reproduction) in the presence of full siblings rather than in the presence of half-siblings, simply because full siblings provide much higher inclusive fitness benefits [19]. The same explanation can be applied to our results showing a consistent association between earlier age at menarche and having no sibling. If you have no siblings, you cannot get any inclusive fitness benefit from helping your mother with them and, thus you should have your menarche sooner. In fact, from the point of view of relatedness, having unrelated step-siblings is the same as having no siblings at all. Interestingly, in his original study [19], Smith never explicitly commented on the association between earlier age at menarche and having no siblings even though, he too reported this association as significant in his final adjusted model.
On the other hand, it is also possible to imagine other potential explanations of the association between sibling relatedness and age at menarche that do not consider indirect fitness benefits at all. It has been previously described that internationally adopted children have up to 20 times higher risk of precocious puberty [29]. Accordingly, results from a study of 396 evacuees from Helsinki sent by their parents to temporary foster families in Denmark and Sweden during the Winter War (1939–1940) indicate that girls from this population sample had earlier menarche [30]. Another study reported that various adverse childhood experiences are also associated with earlier menarche [31]. It is thus conceivable that since ‘gaining’ step-siblings is often connected with moving from one place to another and may call for the development of new social connections (i.e. with step-siblings and step-parents), the associated stress itself may drive earlier menarche. Furthermore, given that full siblings were found to remain closer in adulthood than half-siblings who were raised together like full siblings [32], it seems possible that even during childhood full siblings may provide more psychosocial support to each other than half-siblings. From this point of view, it is the stress itself and not relatedness with siblings or father absence that serves as a cue for a future environment where it is beneficial to reproduce early. One possible benefit of this explanation over invoking inclusive fitness benefits is that it may also explain the difference in age at menarche between girls with no siblings and girls with half/step-siblings. It is conceivable that from a long-term perspective having no siblings is more stressful than having full siblings but less stressful than having half/step-siblings. However, we would like to point out that, as is often a case in biology it is even possible that these alternative explanations (based on indirect fitness benefits or stress) may be right simultaneously.
Overall, our study provides further evidence that sibling relatedness rather than father absence influences age at menarche. Furthermore, it also shows that, in the similarly constructed ALSPAC and Czech ELSPAC cohorts, age at menarche is virtually the same despite the different socioeconomic situation in the two countries.
Supplementary Material
Supplementary Material
Supplementary Material
Supplementary Material
Ethics
Ethical approval for the study was obtained from the ELSPAC Ethics Committee and local research ethics committees. Written informed consent was obtained from all study participants. Project number is ELSPAC/EK/1/2014.
Data accessibility
Data used in this submission are available upon request to the project manager (elspac@recetox.muni.cz). However, restrictions apply to the availability of these data, which were used under licence for the current study, and so are not publicly available. The process of requesting data is described at http://www.elspac.cz/index-en.php?pg=professionals–partnership-establishment.
Authors' contributions
P.L. interpreted the results, participated in the study design and drafted the manuscript. F.Z. carried out the statistical analysis and participated in interpretation. L.K. and J.J. managed the database with cohort data. J.BV. conceived the study, supervised the study and critically revised the manuscript. All authors contributed to the preparation of the manuscript and gave final approval for publication, and agree to be held accountable for the work described therein.
Competing interests
We have no competing interests
Funding
This study was funded by the Ministry of Education, Youth and Sports of the Czech Republic and European Structural and Investment Funds (CETOCOEN PLUS project: CZ.02.1.01/0.0/0.0/15_003/0000469 and the RECETOX research infrastructure: LM2015051 and CZ.02.1.01/0.0/0.0/16_013/0001761). Furthermore, Peter Lenart received support from Brno Ph.D. Talent.
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Associated Data
This section collects any data citations, data availability statements, or supplementary materials included in this article.
Supplementary Materials
Data Availability Statement
Data used in this submission are available upon request to the project manager (elspac@recetox.muni.cz). However, restrictions apply to the availability of these data, which were used under licence for the current study, and so are not publicly available. The process of requesting data is described at http://www.elspac.cz/index-en.php?pg=professionals–partnership-establishment.