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Keywords:

  • developmental instability;
  • first-cousin marriage;
  • inbreeding depression;
  • birth weight

ABSTRACT

  1. Top of page
  2. ABSTRACT
  3. MATERIAL AND METHODS
  4. RESULTS
  5. DISCUSSION
  6. ACKNOWLEDGMENTS
  7. LITERATURE CITED

Historically, medical concerns about the deleterious effects of closely inbred marriages have focused on the risk posed by recessive Mendelian disease, with much less attention to developmental instability. We studied the effects of inbreeding (first-cousin marriage) on growth and fluctuating asymmetry of 200 full-term infants (101 inbred and 99 outbred) whose parents were of similar socioeconomic status in Sivas Province, Turkey. In addition to differences in their mean inbreeding coefficients (f = 1/16 for first cousins and f < 1/1,024 for unrelated parents), the consanguineous parents were less well educated (3 years, on average for both husbands and wives). We measured weight, height, head circumference, and chest circumference of the newborns, as well as four bilateral traits (ear width, ear length, and second and fourth digit lengths). After taking education into account, none of the measures of size (weight, height, head circumference, and chest circumference) and fluctuating asymmetry differed between the inbred and outbred groups. Male children of well-educated parents, however, were larger and had less fluctuating asymmetry. Female children of well-educated parents weighed more than those of less well-educated parents, but were otherwise indistinguishable for height, head circumference, chest circumference, and fluctuating asymmetry. We conclude that inbreeding depression causes neither an increase in fluctuating asymmetry of full-term newborns, nor a decrease in body size. Unmeasured variables correlated with education appear to have an effect on fluctuating asymmetry and size of male children and only a weak effect on size (weight) of female children. Am J Phys Anthropol 153:45–51, 2014. © 2013 Wiley Periodicals, Inc.

Inbreeding, the mating of close relatives, reduces the survival and reproductive fitness of normally outbred plants and animals, including humans (Crow and Kimura, 1970; Lynch and Walsh, 1998; Charlesworth and Willis, 2009). Inbreeding exposes deleterious recessive alleles and reduces heterozygosity. The inbreeding depression that results from such consanguineous mating is mostly a consequence of deleterious recessive alleles in the homozygous state (Charlesworth and Willis, 2009). Nevertheless, one cannot discount true overdominance (heterozygote superiority) at a few critical loci, such as those in the major histocompatibility complex (Kekäläinen et al., 2009; Lie et al., 2009; Oliver et al., 2009). Despite the huge literature on inbreeding depression, it is still unclear whether or not inbreeding has a consistent impact on fluctuating asymmetry, a measure of the stability of development under constant environmental and genetic conditions (Zakharov, 1992). Several articles have shown that inbred lines of various animals have greater fluctuating asymmetry than corresponding outbred lines (Mather, 1953; Beardmore, 1960; van Noordwijk and Scharloo, 1981; Leamy, 1984; Alados et al., 1995; Gomendio et al., 2000; Carter et al., 2009). Other studies, though, have shown no effect of inbreeding on fluctuating asymmetry (Keller and Passera, 1993; Fowler and Whitlock, 1994; Gilligan et al., 2000; Hosken et al., 2000; Leamy et al., 2001; Kruuk et al., 2003). In studies of humans, some small, inbred populations have high fluctuating asymmetry (Bailit et al., 1970; Hershkovitz et al., 1993; Markow and Martin, 1993; Schaefer et al., 2006). In addition, Özener (2010) has shown that a group of Turkish high school students whose parents are first cousins weigh less than, and have greater fluctuating asymmetry than, a matched group whose parents are not closely related. Here, we show that inbred human infants of first-cousin marriages do not have greater fluctuating asymmetry than those of outbred ones. Moreover, their size (weight, height, head circumference, and chest circumference) is not reduced. In addition, the parents of the inbred children are less well educated than the parents of the outbred children, which almost certainly confounds fluctuating asymmetry and size with behaviors associated with education. Male children of less well educated parents were smaller and more asymmetrical at birth than those of more educated parents.

Fluctuating asymmetry is the extent to which the average individual departs from perfect symmetry, be it bilateral, radial, or translatory symmetry. It is an indicator of developmental instability and random developmental noise. A large literature suggests that various forms of genetic and environmental stress can increase fluctuating asymmetry, though the patterns are inconsistent (Palmer and Strobeck, 1986; Markow, 1994; Møller and Swaddle, 1997; van Dongen, 2006; Graham et al., 2010). In addition, the genetic basis of fluctuating asymmetry is unclear (Leamy and Klingenberg, 2005), but the resilience of gene regulatory, metabolic, and protein-protein interaction networks operating in noisy developmental systems appears to play a role (Graham et al., 2012; Levy and Siegal, 2008).

MATERIAL AND METHODS

  1. Top of page
  2. ABSTRACT
  3. MATERIAL AND METHODS
  4. RESULTS
  5. DISCUSSION
  6. ACKNOWLEDGMENTS
  7. LITERATURE CITED

Sample

This study was carried out in Sivas Province of the Central Anatolian Region of Turkey. This province, with a population of ∼300,000, is ranked 49th of 81 provinces according to the Survey of Socio-Economic Development Rankings of the Provinces and Regions of Turkey (issued in 2012 by the Ministry of Development). While 40% of the population makes a living from agriculture and livestock breeding in rural areas, the inbreeding rate in this province is 25% according to the Turkish Statistical Institute (http://www.turkstat.gov.tr/). Our study was conducted in the New Born Clinic of Cumhuriyet University Hospital, which is the only state University in this Province. Mostly lower/medium socio-economic families apply to this clinic, where five to six deliveries occur per day.

The study group consisted of full-term newborns with gestational ages between 37 and 41 weeks. There was no difference in mean gestational age between inbred and outbred groups (F1,198 = 0.90, P = 0.35). After deliveries in the new-born clinic, babies were monitored for 2–3 days. We observed 200 new-born babies and their parents (101 inbred and 99 outbred) in the clinic where healthy babies were hospitalized. The most common type of inbreeding in Turkey is among first cousins (Tunçbilek and Koc, 1994; Koc, 2008). Inbred group parents were selected amongst first degree cousins (f = 1/16). The outbred group parents were selected from unrelated parents, excluding even third-degree (f = 1/256) and fourth-degree cousins (f = 1/1,024). Prior to measurement of the infants, parents were asked to voluntarily fill informed consent forms approved by the Ethics Committee of Cumhuriyet University.

Measurements

The parents reported their level of education, number of experienced pregnancies, and their degree of genetic relatedness. Physical measurements on the infants were performed in the clinic where new-born babies were hospitalized. Bilateral measurements were taken when the babies were asleep. Manning et al. (2010) concluded that direct measurement of finger length is as reliable as indirect measurements (i.e., photocopy). Lengths of the second and fourth digits were measured from the crease proximal to the palm to the tip of the finger. Ear length and width were measured in accordance with the techniques proposed by the International Biological Program (Weiner and Lourie, 1969). All measurements were taken to the nearest 0.01 mm using a Vernier caliper.

Bilateral data were obtained using the blind measurement technique (Palmer and Strobeck, 2003). First, measurements from the right side of the body were taken in the order listed above. Then data were collected from the left side in the same manner. For the second set of measurements this exact procedure was repeated without referring to the prior data. There was a gap of ∼15 min between the two sets of measurements. The person making the measurements was unaware of the marriage status of the parents.

Birth weight with naked newborns in supine position was obtained soon after birth by digital scale with 10-g sensitivity. The height was measured in supine position with full extension of knee and distance between top of head and heel when pressed against a vertical surface and rolled on a stabilizing board by infantometer to the nearest 0.1 cm. Circumferences were measured by nonextendable measuring tape to the nearest 0.1 cm. Head circumference was obtained by placing tape along the largest occipitofrontal diameter along and over the occiput and eyebrow. The chest circumference was measured by placing measuring tape along the point of the nipples.

Measurement error of bilateral measurements

Of the 200 infants in this study, 60 (30 males, 30 females) were measured twice to estimate measurement error. A mixed-model ANOVA was used to estimate repeatability of the asymmetry. In this method, the main fixed effect is sides (S; right or left). The block effect is individuals (I). The sides by individual interaction (S × I) is a mixed effect. An error term (m) represents measurement error (replications within sides by individuals). The ratio of the S × I mean square to the residual (error) mean square provides an F-test of whether the estimated nondirectional asymmetry (i.e., fluctuating asymmetry and antisymmetry) is significantly greater than zero (Palmer and Strobeck, 2003). The variance component for the S × I interaction is sometimes used to estimate fluctuating asymmetry and its statistical significance is determined by an F statistic testing the null hypothesis inline image (Graham et al., 2010). The variance components of the side-by-individual interaction terms were significantly greater than zero for all four traits (Ear length: F1,59 = 19.82, P < 0.001; ear width: F1,59 = 17.46, P < 0.001; second digit length: F1,59 = 11.32, P < 0.001; fourth digit length: F1,59 = 12.39, P < 0.001). Measurement error accounted for only 0.863–1.416 percent of the total variance and 10.63–17.55 percent of the asymmetry variance (S × I interaction).

Trait-size dependence, directional asymmetry, and normality

Tests for size-dependence are best conducted before testing for ideal fluctuating asymmetry, because size-dependent heterogeneity in asymmetry variation can yield leptokurtosis in the frequency distributions of d = RL, obscure subtle antisymmetry, or generate spurious differences among groups (see Palmer and Strobeck, 2003). Spearman's rank correlation coefficient (rs) was used to quantify the relationships between absolute asymmetry |d| = |RL| and mean trait size (R + L)/2 for all traits, according to group (marriage type and sex). There was no indication of size dependency for any of these traits (all rs ≤ |0.124| and all P ≥ 0.224). To detect the existence of directional asymmetry within the groups, one-sample t-tests were used. The null hypothesis for this test is that the mean signed asymmetry (d) equals zero. None of the traits were found to be directionally asymmetric (all t ≤ 1.628 and all P ≥ 0.110).

To determine the departures from normality, we calculated skewness and kurtosis of d (i.e., signed asymmetries) for all traits (Sokal and Rohlf, 1995). In the inbred group, ear lengths were positively skewed (P < 0.05); in the outbred group, second and fourth digit lengths were negatively skewed (P < 0.05). In the inbred group, ear widths (P < 0.05) and in the outbred group second digit lengths (P < 0.01) had leptokurtic distributions. After log transformation, both log R − log L and log(RL), a number of measurements were still significantly leptokurtotic and skewed. Possible outliers were identified visually from scatter plots as suggested by Palmer and Strobeck (2003), and then these measures were tested and removed according to Grubbs' test in the raw and signed asymmetry data. Seven measurements in the inbred group and 11 measurements in the outbred group were found to be outliers and therefore excluded from the sample. A normal distribution was attained after 18 measurements were excluded from the data set.

Fluctuating asymmetry analysis

Signed asymmetries (d) and absolute asymmetries |d| were determined for each individual. For composite fluctuating asymmetry (CFA), we summed the standardized absolute asymmetries of the four traits (Leung et al., 2000); the asymmetry of each trait was standardized by its mean fluctuating asymmetry: inline image, where inline image is the fluctuating asymmetry of trait i, averaged across all individuals in the study.

The composite index of fluctuating asymmetry is most effective when the unsigned asymmetries of the component traits are all responding to stress in the same way. Hence, the unsigned asymmetries of the four bilateral traits should be positively correlated. Nevertheless, these same trait asymmetries should be independent, in the sense that they do not form a developmental unit. Therefore, the signed asymmetries should be uncorrelated (Klingenberg et al., 2001). To evaluate these underlying assumptions of the composite index of fluctuating asymmetry, we examined the correlations of unsigned and signed asymmetries of the entire data set, and also for the four subgroups (inbred females, outbred females, inbred males, outbred males). We also looked at Kendall's Coefficient of Concordance (W) for evidence of an individual asymmetry parameter.

Multivariate and univariate analyses of covariance

To test the hypothesis that children of first cousins had greater fluctuating asymmetry and were smaller at birth than those of unrelated couples, we used a multivariate analysis of covariance, with weight, height, head circumference, chest circumference, and CFA as the dependent variables and marriage type (inbred, outbred) as the independent variable, with total years of parental education as the covariate. We did separate analyses for male and female children, because a marriage type by sex analysis is an example of pseudofactorialism (Hurlbert, 2013) and because the slopes of most of the dependent variables regressed on total years of education were not homogeneous for sex. Variances of the dependent variables and slopes of the regression lines were all homogeneous.

Because this is an observational study, and individuals are not randomly assigned to treatment groups, it is essential to test the assumption that covariate and treatment are independent (Huitema, 1980). In all cases, total years of education and marriage type were not independent, which complicates interpretation.

RESULTS

  1. Top of page
  2. ABSTRACT
  3. MATERIAL AND METHODS
  4. RESULTS
  5. DISCUSSION
  6. ACKNOWLEDGMENTS
  7. LITERATURE CITED

With the exception of the correlations of ear length and width asymmetries with second digit asymmetry (r ≤ 0.091, P ≥ 0.199), the unsigned asymmetries of the four bilateral traits were positively correlated across the entire data set (r ≥ 0.173, P ≤ 0.015). Moreover, Kendall's Coefficient of Concordance (W) was 0.089 (X2 = 53.174, df = 3, P < 0.001) for the entire data set. The same correlations in the four marriage type by sex subpopulations were much less consistent. Nevertheless, three of the four subgroups (inbred and outbred females and inbred males) had significant values of W (W ≥ 0.090, df = 3, P ≤ 0.003) so there was evidence for an individual asymmetry parameter in these groups. In addition, the level of integration among the four bilateral traits was low, even for ear length and width and digits 2 and 4. Among the 24 correlations among these four variables (signed asymmetry), subdivided into four combinations of marriage type and sex, only four correlations were statistically significant (P < 0.05). Of these four correlations, one was negative, the other three were moderately positive. Moreover, the significant correlations were inconsistent with respect to marriage type and sex. Consequently these four traits are reasonably independent.

Descriptive statistics for inbred and outbred newborns are presented in Tables 1 (for males) and 2 (for females). Both male and female parents of outbred infants had, on average, 3.1 years more education than those of the inbred infants. Male parents had 1.36 years more education than female parents. The differences in education (for both fathers and mothers) between marriage types were highly significant (MANOVA, Wilk's λ = 0.815, F2,197 = 22.39, P < 0.001). Moreover, the number of years of education of male and female parents were highly correlated for both inbred (rpearson = 0.692, n = 101, P < 0.001) and outbred (rpearson = 0.816, n = 99, P < 0.001) marriages. After taking total years of education into account (ANCOVA, F1,197 = 30.30, P < 0.001), the total number of pregnancies did not differ between consanguineous and nonconsanguineous couples (ANCOVA, F1,197 = 0.359, P > 0.540).

Table 1. Descriptive statistics (means and standard deviations) for inbred and outbred males
 Inbred malesOutbred males
MeanSDMeanSD
  1. a

    Means are adjusted after taking total years of education into account.

Mother's ed. level (year)5.982.539.043.69
Father's ed. level (year)7.353.2110.453.93
Weight (kg)a3.405.563.356.42
Height (cm)a50.571.9450.351.81
Head circumference (cm)a35.490.9935.150.97
Chest circumference (cm)a33.821.0233.411.20
Ear length FA (mm)a1.311.501.161.12
Ear width FA (mm)a1.261.040.951.06
Second digit FA (mm)a0.530.530.610.68
Fourth digit FA (mm)a0.570.650.560.48
Composite FA (mm)a0.920.500.820.56
Table 2. Descriptive statistics (means and standard deviations) for inbred and outbred females
 Inbred femalesOutbred females
MeanSDMeanSD
  1. a

    Means are adjusted after taking total years of education into account.

Mother's ed. level (year)6.042.799.003.89
Father's ed. level (year)7.203.3210.484.51
Weight (kg)a3.174.503.245.52
Height (cm)a50.101.3350.181.56
Head circumference (cm)a34.841.0035.160.83
Chest circumference (cm)a33.250.8533.561.03
Ear length FA (mm)a1.241.670.991.09
Ear width FA (mm)a1.000.940.881.07
Second digit FA (mm)a0.770.650.440.76
Fourth digit FA (mm)a0.380.580.410.45
Composite FA (mm)a0.850.630.680.54

We used a multivariate analysis of covariance (MANCOVA) to test the null hypothesis that the mean vectors of inbred and outbred newborns were sampled from the same population. Total years of parental education was the covariate. After removing the highly significant effect of education, first-cousin marriages had no detectable influence on the mean vectors of male and female newborns (Table 3).

Table 3. Multivariate analyses of covariance for male and female newborns
SexSource of variationHypothesis DFError DFWilk's λFP
  1. The dependent variables are weight, height, head circumference, chest circumference, and composite fluctuating asymmetry. The covariate is total years of education.

MaleMarriage type5920.9520.9320.464
Total years of education5920.7555.959<0.001
FemaleMarriage type5930.9431.1340.348
Total years of education5930.8533.1930.011

As anticipated from the results of the MANCOVAs, first-cousin marriages had no detectable influence on any of the five dependent variables, in either males or females (Table 4). In contrast, all of the dependent variables in male newborns varied with total years of education, whereas only one dependent variable in females (weight) did so. More educated parents had larger male newborns (greater weight, taller stature, and larger head and chest circumferences) having less fluctuating asymmetry (Fig. 1). More educated parents had female newborns that weighed more.

image

Figure 1. Linear regression of CFA on total years of education, by sex and marriage type. Open circles and dashed lines are offspring of first cousins. Closed circles and solid lines are offspring of unrelated couples.

Download figure to PowerPoint

Table 4. Univariate analyses of covariance for male and female newborns
SexDependent variableMarriage typeTotal years of educationError
DFMSFPDFMSFPDFMS
  1. The independent variable is marriage type (inbred, outbred) and the covariate is total years of education (male parent plus female parent).

MaleWeight137.710.110.74211,595.14.610.0349634.6
Height15.431.700.196136.011.30.001963.20
Head circumference12.312.620.10918.219.300.003960.88
Chest circumference13.462.980.08817.696.620.012961.16
Composite FA10.170.730.39514.7520.1<0.001960.24
FemaleWeight1107.10.440.50711,430.55.920.01797241.6
Height10.110.0520.82011.0760.510.477972.11
Head circumference12.212.590.11110.0410.050.828970.85
Chest circumference11.952.180.14310.650.720.397970.90
Composite FA10.571.640.20310.371.060.306970.35

DISCUSSION

  1. Top of page
  2. ABSTRACT
  3. MATERIAL AND METHODS
  4. RESULTS
  5. DISCUSSION
  6. ACKNOWLEDGMENTS
  7. LITERATURE CITED

When total years of education were taken into account as a covariate, inbred (f = 1/16) Turkish newborns in Sivas did not have greater fluctuating asymmetry than unrelated (f < 1/1,024) newborns. Moreover, they were of normal size at birth. Özener (2010), in contrast, found that inbred teens from urban Ankara had greater composite fluctuating asymmetry and were, on average, 3.52 kg lighter and 3.17 cm shorter than comparable outbred teens. While several previous studies have failed to detect greater fluctuating asymmetry of inbred populations (for reviews, see Leamy and Klingenberg, 2005; Graham et al., 2010), smaller body size is commonly observed in studies of inbreeding depression (Bittles and Black, 2010b) and the lack of it here requires an explanation.

The normal birth weight among the inbred infants may arise for several reasons. First, low weight may be an effect of inbreeding seen later in life. A more likely explanation, however, is that our study only examined full-term infants, with gestational ages between 37 and 41 weeks. We did not examine preterm infants, which almost certainly would have had lower birth weight (Kramer et al., 2001), and perhaps greater fluctuating asymmetry as well. In Jordan, preterm delivery accounts for most of the differences in birth weight between inbred (19.9% preterm) and outbred (12.3% preterm) infants (Obeidat et al., 2010). Similar results were obtained in Lebanon (Mumtaz et al., 2007; Mumtaz et al., 2010).

Because ours is an observational study, we cannot assign cause and effect with certainty. There may have been other, unobserved, variables influencing fluctuating asymmetry. For example, the parents of the inbred infants in our study had about 3 years less education than those of the outbred infants. Education may be confounded with other variables that influence size and asymmetry. If two populations in an observational study differ on one covariate (education), they must differ on others (Huitema, 1980). Smoking rates, for example, are greater among less educated individuals (Giskes et al., 2005). In Sivas, 4.9% of high school girls and 36.5% of high school boys smoke, and the percentage is about 10% higher for students whose parents have less than a high school education for fathers and less than a primary school education for mothers (Golbasi et al., 2011). Although education itself is probably causally unrelated to fluctuating asymmetry, it is almost certainly confounded with smoking and diet. Kieser et al. (1997), for example, have shown that dental asymmetry in humans is related to both smoking and obesity. Finally, when Obeidat et al. (2010) took smoking and other variables (i.e., age, body mass index, income, residency, smoking, parity, family history of low birth weight, preterm delivery, congenital anomalies, and medical problems during pregnancy) into account in a very large study of inbreeding in a Jordanian population (n = 3,269), some of the negative effects, such as low birth weight and increased number of still births, were no longer significant. Preterm deliveries and congenital malformations, however, were still higher in the inbred Jordanian group, even when smoking was taken into account.

Even if smoking and diet play no role in the relationship between consanguineous mating, education, size at birth, and fluctuating asymmetry, it is also possible that our results are an artifact of another kind of nonrandom mate selection. Lynch and Walsh (1998) have suggested that, in humans, low-quality (or low social status) parents may be more likely to marry other low-status individuals who may, or may not, be close relatives, simply because their range of prospective mates is limited. Denic and Agarwal (2012) describe just this scenario in a family that had lost social status for political reasons. If this is the case, then the observed inbreeding depression may be exaggerated by the expression of alleles (not necessarily recessive ones) associated with the low fitness. If low-status individuals are less likely to extend their education, and if low status has a genetic component, then the relationship between education, size at birth, and fluctuating asymmetry could also have a genetic component. This argument, however, makes more sense for societies in which first-cousin marriages are rarer than they are in central Turkey. Throughout Turkey, nearly 9% of the richest households feature first-cousin marriages (Koc, 2008). While this rate is less than half that of the poorest households, which have a rate of 22.3% in Turkey, it is still high by European standards. This then begs the question, why is inbreeding so prevalent in some societies and not in others?

Under some circumstances, consanguineous marriages may produce more children than nonconsanguineous pairings (Helgason et al., 2008; Bittles and Black, 2010a). And entire populations can benefit from consanguineous mating in places where malaria is endemic, by exposing alleles such as those associated with α+-thalassemia to positive selection (Denic et al., 2008, 2013; but see Bittles, 2011). This is not a new idea, however. Sewell Wright (1932, 1982) long argued that mutation, inbreeding, and selection interact to increase a population's rate of adaptation. This may be the situation in Turkey. Until recently, malaria was endemic in southeastern Turkey (Özbilgin et al., 2011; Piyal et al., 2013). This is an area where consanguineous marriages are common.

In conclusion, first cousin marriages in Turkey do not appear to increase the level of fluctuating asymmetry in newborns (this article), though they do so in urban teens (Özener, 2010). Moreover, first cousin marriages in Sivas do not cause size reduction of newborns, but this may be an artifact of selecting only full-term infants in our study. Future studies should also look into correlates of parental education, such as smoking behavior, diet, and exposure to environmental toxins.

ACKNOWLEDGMENTS

  1. Top of page
  2. ABSTRACT
  3. MATERIAL AND METHODS
  4. RESULTS
  5. DISCUSSION
  6. ACKNOWLEDGMENTS
  7. LITERATURE CITED

The authors thank Aslihan Hamamcioglu for helping with the data collection and Catherine Chamberlin-Graham for helping with the literature search. They also thank the New Born Clinic of Cumhuriyet University Hospital for making this research project possible, and the families who participated in the study. Two reviewers provided helpful comments.

LITERATURE CITED

  1. Top of page
  2. ABSTRACT
  3. MATERIAL AND METHODS
  4. RESULTS
  5. DISCUSSION
  6. ACKNOWLEDGMENTS
  7. LITERATURE CITED
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