Behavioral genetic research designs have often been attacked because they rely on comparing monozygotic twins (MZ) to dizygotic twins (DZ). Critics of twin-based research maintain that MZ twins look m

In this study, our aim is twofold. First, we analyze empirical esti - mates of the relative contributions of genes and environment for virtually all human traits investigated in the past 50 years. Second, we assess empirical evidence for the presence and relative importance of non-additive genetic influences on all human traits studied. We rely on classical twin studies, as the twin design has been used widely to disentangle the relative contributions of genes and environment, across a variety of human traits. The classical twin design is based on contrasting the trait resemblance of monozygotic and dizygotic twin pairs. Monozygotic twins are genetically identical, and dizygotic twins are genetically full siblings. We show that, for a majority of traits (69%), the observed statistics are consistent with a simple and parsi - monious model where the observed variation is solely due to additive genetic variation. The data are inconsistent with a substantial influence from shared environment or non-additive genetic variation. We also show that estimates of heritability cluster strongly within functional domains, and across all traits the reported heritability is 49%. Our results are based on a meta-analysis of twin correlations and reported variance components for 17,804 traits from 2,748 publications includ - ing 14,558,903 partly dependent twin pairs, virtually all twin studies of complex traits published between 1958 and 2012. This study provides the most comprehensive analysis of the causes of individual differences in human traits thus far and will guide future gene-mapping efforts. All results can be visualized with the accompanying MaTCH webtool.

RESULTS The distribution of studied traits is nonrandom We systematically retrieved published classical twin studies in which observed variation in human traits was partitioned into genetic and environmental influences. For each study, we collected reported Meta-analysis of the heritability of human traits based on fifty years of twin studies Tinca J C Polderman 1 ,10 , Beben Benyamin 2 ,10 , Christiaan A de Leeuw 1 ,3 , Patrick F Sullivan 4 – 6 , Arjen van Bochoven 7, Peter M Visscher 2 ,8 ,11 & Danielle Posthuma 1 ,9 ,11 1Department of Complex Trait Genetics, VU University, Center for Neurogenomics and Cognitive Research, Amsterdam, the Netherlands. 2Queensland Brain Institute, University of Queensland, Brisbane, Queensland, Australia. 3Institute for Computing and Information Sciences, Radboud University Nijmegen, Nijmegen, the Netherlands. 4Center for Psychiatric Genomics, Department of Genetics, University of North Carolina, Chapel Hill, North Carolina, USA. 5Department of Psychiatry, University of North Carolina, Chapel Hill, North Carolina, USA. 6Department of Medical Epidemiology and Biostatistics, Karolinska Institutet, Stockholm, Sweden. 7Faculty of Sciences, VU University, Amsterdam, the Netherlands. 8University of Queensland Diamantina Institute, Translational Research Institute, Brisbane, Queensland, Australia. 9Department of Clinical Genetics, VU University Medical Center, Neuroscience Campus Amsterdam, Amsterdam, the Netherlands. 10These authors contributed equally to this work. 11These authors jointly supervised this work. Correspondence should be addressed to D.P. ( [email protected]).

Received 13 February; accepted 1 April; published online 18 May 2015; doi:10.1038/ng.328 5 Insight into the nature of observed variation in human traits is impor - tant in medicine, psychology, social sciences and evolutionary biology. It has gained new relevance with both the ability to map genes for human traits and the availability of large, collaborative data sets to do so on an extensive and comprehensive scale. Individual differences in human traits have been studied for more than a century, yet the causes of variation in human traits remain uncertain and controversial. npg © 2015 Nature America, Inc. All rights reserved. Nat u r e G eNe t i c s VO L U M E 4 7 | N U M B E R 7 | J U LY 2 0 1 5 7 0 3 twin correlations for continuous traits and contingency tables for dichotomous traits, estimates from genetic model-fitting and study characteristics (sample size, population, age cohort and ascertain - ment scheme) ( Supplementary Table 1 and Supplementary Note ). We manually classified the investigated traits using the chapter and subchapter levels of the International Classification of Functioning, Disability and Health (ICF) or the International Statistical Classification of Diseases and Related Health Problems (ICD- 10) (Online Methods). ICD-10 and ICF subchapter levels refer to actual diseases (for example, atopic dermatitis) and traits (for example, temperament and personality functions), respectively. We identified 2,748 relevant twin studies, published between 1958 and 2012. Half of these were published after 2004, with sample sizes per study in 2012 of around 1,000 twin pairs ( Supplementary Table 2). Each study could report on multiple traits measured in one or several samples. These 2,748 studies reported on 17,804 traits. Twin subjects came from 39 different countries, with a large proportion of studies (34%) based on US twin samples. The continents of South America (0.5%), Africa (0.2%) and Asia (5%) were heavily under- represented ( Fig. 1a,b and Supplementary Table 3 ). The total number of studied twins was 14,558,903 partly dependent pairs, or 2,247,128 when correcting for reporting on multiple traits per study. The majority of studies (59%) were based on the adult population (aged 18–64 years), although the sample sizes available for studies of the elderly population (aged 65 years or older) were the largest ( Supplementary Table 4). Authorship network analyses showed that 61 communities of authors wrote the 2,748 published studies. The 11 largest authorship communities contained >65 authors and could be mapped back to the main international twin registries, such as the Vietnam Era Twin Registry, the Finnish Twin Cohort and the Swedish Twin Registry ( Supplementary Fig. 1 ).

The investigated traits fell into 28 general trait domains. The dis - tribution of the traits evaluated in twin studies was highly skewed, with 51% of studies focusing on traits classified under the psychi - atric, metabolic and cognitive domains, whereas traits classified under the developmental , connective tissue and infection domains together accounted for less than 1% of all investigated traits ( Fig. 1c and Supplementary Tables 5 –7). The ten most investigated traits were temperament and personality functions, weight maintenance functions, general metabolic functions, depressive episode, higher- level cognitive functions, conduct disorders, mental and behavioral disorders due to use of alcohol, anxiety disorders, height and mental and behavioral disorders due to use of tobacco. Collectively, these traits accounted for 59% of all investigated traits. 0 200 400 600 800 1,000 1,200 Connective tissue Developmental InfectionAging Mortality Neoplasms Cell Hematological Social values Muscular Gastrointestinal Nutritional Dermatological Social interactions Ear, nose, throat Immunological Activities Ophthalmological Respiratory Reproduction Endocrine Environment Cardiovascular Skeletal Neurological Cognitive Metabolic Psychiatric a b c d Studies (n)Continuous 89% Dichotomous 10% Dichotomous, ascertained 1% Other, non- disease75% Disease 8%Symptoms of disease17% Number of investigated traits Average number of twin pairs per study 1 2 3 4 –0.5 0 0.5 1.0 rMZ R2 = 0.009468 1 2 3 4 –0.5 0 0.5 1.0 rDZ R2 = 0.0003032 1.0 1.5 2.0 2.5 3.0 3.5 4.0 4.5 0 0.2 0.4 0.6 0.8 1.0 log 10 (n pairs) h2 R2 = 0.001651 1.0 1.5 2.0 2.5 3.0 3.5 4.0 4.5 0 0.2 0.4 0.6 0.8 log 10 (n pairs) c2 R2 = 0.005001 1 6,74610 2,104 Figure 1 Distribution of the investigated traits in virtually all twin studies published between 1958 and 2012. ( a) The number of investigated traits in classical twin studies across all countries. ( b) The average number of twin pairs included per study across countries. ( c) The number of investigated traits according to functional trait domain and trait characteristic (inset). ( d) Monozygotic and dizygotic twin correlations and reported estimates of h 2 and c 2 as a function of sample size. Contour lines indicate the density of the data in that region. The lines are colored by ‘heat’ from blue to red, indicating increasing data density. A n Al y s i s npg © 2015 Nature America, Inc. All rights reserved. 7 0 4 V O L U M E 4 7 | N U M B E R 7 | J U LY 2 0 1 5 Nat u r e G e Ne t i c s A nAl y s i s Equal contribution of genes and environment We did not find evidence of systematic publication bias as a function of sample size (for example, where studies based on relatively small sam - ples were only published when larger effects were reported) ( Fig. 1d, Supplementary Figs. 2 –6 and Supplementary Tables 8 –11 ). We cal - culated the weighted averages of correlations for monozygotic ( r MZ ) and dizygotic ( r DZ) twins and of the reported estimates of the relative contributions of genetic and environmental influences to the investi - gated traits using a random-effects meta-analytic model to allow for heterogeneity across different studies ( Supplementary Tables 12–15).

The meta-analyses of all traits yielded an average r MZ of 0.636 (s.e.m. = 0.002) and an average r DZ of 0.339 (s.e.m. = 0.003). The reported heritability ( h 2) across all traits was 0.488 (s.e.m. = 0.004), and the reported estimate of shared environmental effects ( c 2) was 0.174 (s.e.m. = 0.004) ( Fig. 2a,b , Table 1 and Supplementary Fig. 7 ). Estimates of h2 and c2 cluster in functional domains We found that heritability estimates clustered in functional domains, with the largest heritability estimates for traits classified under the ophthalmological domain ( h 2 = 0.712, s.e.m. = 0.041), followed by the ear, nose and throat ( h 2 = 0.637, s.e.m. = 0.064), dermatological ( h 2 = 0.604, s.e.m. = 0.043) and skeletal ( h 2 = 0.591, s.e.m. = 0.018) domains. The lowest heritability estimates were for traits in the environment, reproduction and social values domains ( Fig. 2d and Supplementar y Table 16 ). All weighted averages of h 2 across >500 distinct traits had a mean greater than zero ( Supplementary Tables 17 –24 ). The lowest reported heritability for a specific trait was for gene expression, with an estimated h 2 = 0.055 (s.e.m. = 0.026) and an estimated c 2 of 0.736 (s.e.m. = 0.033) (but note that these trait aver - ages are based on reported estimates of variance components derived from only 20 data points reporting on the expression levels of 20 genes; Supplementary Table 21 ). We found the largest influence of c 2 for traits in the cell domain ( c 2 = 0.674, s.e.m. = 0.048), followed by traits in the infection ( c 2 = 0.351, s.e.m. = 0.153), hematological ( c 2 = 0.324, s.e.m. = 0.090), endocrine ( c 2 = 0.322, s.e.m. = 0.050), reproduction ( c 2 = 0.320, s.e.m. = 0.061), social values ( c 2 = 0.271, s.e.m. = 0.032), environment ( c 2 = 0.269, s.e.m. = 0.020) and skeletal ( c 2 = 0.265, s.e.m. = 0.019) domains ( Fig. 2d and Supplementary Table 16 ). 0 0.2 0.4 0.6 0.8 1.0 0 0.2 0.4 0.6 0.8 1.0 All traits Skeletal Reproduction Ear, nose, throat Metabolic Ophthalmological Dermatological Respiratory Neurological CognitiveActivities Cardiovascular Endocrine Psychiatric Environment Gastrointestinal Social values Nutritional Social interactions d rMZ rMZM rMZF rDZ rDZSS rDZM rDZF rDOS h2 h2 SS h2 M h2 F c2 c2 SS c2 M c2 F c2 F Correlation (r) Frequency –0.5 0 0.5 1.00 50 150 250 MZ DZ a 0 0.2 0.4 0.6 0.8 1.0 rMZ rMZM rMZF rDZ rDZSS rDZM rDZF rDOS 0 0.2 0.4 0.6 0.8 1.0 0–11 12–17 18–64 Age (years) 65+ h2 h2 SS h2 M h2 F c2 c2 SS c2 M c –0.5 00.5 1.0 –0.50 0.5 1.0 rMZ rDZ b R2 = 0.4167 Figure 2 Twin correlations and heritabilities for all human traits studied. ( a) Distribution of r MZ and r DZ estimates across the traits investigated in 2,748 twin studies published between 1958 and 2012. r MZ estimates are based on 9,568 traits and 2,563,628 partly dependent twin pairs; r DZ estimates are based on 5,220 traits and 2,606,252 partly dependent twin pairs ( Table 1). ( b) Relationship between r MZ and r DZ, using all 5,185 traits for which both were reported. ( c) Random-effects meta-analytic estimates of twin correlations (top) and reported variance components (bottom) across all traits separately for four age cohorts. Error bars, standard errors. ( d) Random-effects meta-analytic estimates of twin correlations (top) and reported variance components (bottom) across all traits, and within functional domains for which data on all correlations and variance components were available. Error bars, standard errors.

npg © 2015 Nature America, Inc. All rights reserved. Nat u r e G eNe t i c s V O L U M E 4 7 | N U M B E R 7 | J U LY 2 0 1 5 7 0 5 A nAl y s i s Heterogeneity of twin correlations across sex and age Across all traits, the weighted averages of twin correlations and reported h 2 and c 2 values did not show evidence of heterogeneity by sex, although there was some evidence for lower correlation in opposite- sex twin pairs in comparison to same-sex dizygotic twin pairs ( Table 1 and Supplementary Note ). The data showed a decrease in monozygotic and dizygotic twin resemblance after adolescence and an accompanying decrease in the estimates of both h 2 and c 2 (Fig. 2 c and Supplementary Table 15 ).

In the top 20 most investigated traits for twin correlations, the weighted estimates did not show consistent evidence for heterogene - ity by sex, with r MZM (monozygotic male twin correlation) and r MZF (monozygotic female twin correlation) as well as r DZM (dizygotic male twin correlation) and r DZF (dizygotic female twin correlation) remarkably similar across the majority of the top 20 specific traits investigated ( Fig. 3), although for several traits the correlations for opposite-sex twin pairs were lower than the estimates for same-sex twin pairs, mostly after age 11 years (for example, for weight mainte - nance functions, functions of brain, mental and behavioral disorders due to the use of alcohol, and mental and behavioral disorders due to the use of tobacco ). Heterogeneity of weighted twin correlations by age was more prominent than heterogeneity by sex ( Fig. 3). For example, when considering r MZ , we note that for most of the top 20 investigated traits the estimate tended to decrease with age, especially after adolescence, a trend that was generally mirrored in the r DZ estimates ( Fig. 3). Meta-analysis results across all traits across different countries are provided in Supplementary Table 25. Model-fitting and selection leads to underestimation of h2 Falconer’s equations can be used to calculate ĥ 2 and ĉ 2 on the basis of twin correlations 1 8. In these equations, ĥ 2 = 2 × (r MZ − r DZ) and ĉ 2 = 2 × r DZ − r MZ . When these are applied using the weighted averages of r MZ and r DZ, we find an ĥ 2 estimate of 2 × (0.636 − 0.339) = 0.593 and a ĉ 2 estimate of 2 × 0.339 − 0.636 = 0.042 ( Table 1 and Supplementary Fig. 8 ). We note that the ĥ 2 estimate based on twin correlations is larger than the weighted average of reported h 2 values (Supplementary Figs. 9 and 10). As a consequence, the ĉ2 estimate based on twin correlations is lower than the weighted average of the reported c 2 component. To test whether this discrepancy was due to a bias in studies reporting only twin correlations or only variance compo - nents, we conducted the meta-analysis only on studies reporting both.

This analysis yielded similar estimates with a similar discrepancy ( Supplementary Table 13 ), ruling out the explanation that twin cor - relations may have been reported on traits that happened to be more heritable than traits for which the estimates of variance components were reported. Through theory, we show that such a discrepancy can arise when individual studies represent a mixture of traits that follow a pattern of r MZ > 2r DZ and r MZ < 2r DZ and where the choice of fitting a model that includes shared environment or non-additive genetic influences is based on the observed pattern of twin correlations ( Supplementary Note ). More specifically, because c 2 and non-additive genetic influences cannot be estimated simultaneously from twin correlations, an ACE model (for additive genetic (A), common envi - ronmental (C), and error or non-shared environmental (E) influences) is fitted to the data if 2 r DZ − r MZ > 0. In contrast, if 2 r DZ − r MZ < 0, an ADE model, including non-additive genetic (D) instead of com - mon environmental (C) influences, is selected. This leads to sampling bias in estimating h 2 from the full model. We show ( Supplementary Table 26 ) that such per-study choices cause bias and can lead to a 10% downward bias in the reported estimates of h 2 in comparison to those based on twin correlations, consistent with the observed discrepancy between our meta-analysis of variance component estimates calculated from twin correlations and the reported variance components. Overall twin correlations imply a simple additive model There may be many causes of similarities and differences within monozygotic and dizygotic twin pairs, and these are typically inter - preted in terms of additive or non-additive genetic influences and shared or non-shared environmental influences 1 9. Yet, there are essentially only two estimable and testable variance components of interest in the twin design. Therefore, inference from classical twin studies on all underlying, unobserved sources of variation that lead to the resemblance between relatives is limited. However, there are two simple and parsimonious hypotheses that can be tested across traits from estimated correlation coefficients for monozygotic twin pairs ( r MZ ) and dizygotic twin pairs ( r DZ). The first is that the cor- relations for the monozygotic and dizygotic twin populations ( ρMZ and ρDZ) are the same, consistent with twin resemblance being solely due to non-genetic factors. The second hypothesis involves a two - fold ratio of ρMZ to ρDZ, consistent with twin resemblance being solely due to additive genetic variation. Across-trait consistency with either of these hypotheses is not a proof of these simple models but would provide an extremely parsimonious model against which other experimental designs (for example, DNA sequence–based studies) should be tested. For the vast majority of traits (84%), Table 1 Weighted means of twin correlations and variance components across all human traits investigated in a classical twin study and published between 1958 and 2012 Statistic Estimate (s.e.m.)n traitsn pairs r MZ 0.636 (0.002) 9,5682,563,627 r MZM 0.617 (0.004) 4,5181,070,962 r MZF 0.626 (0.004) 4,3601,171,841 r DZ 0.339 (0.003) 5,2202,606,252 r DZSS 0.345 (0.003) 6,1081,752,952 r DZM 0.321 (0.003) 4,4121,039,238 r DZF 0.342 (0.004) 4,2551,068,562 r DOS 0.302 (0.005) 2,342898,610 h 2 0.488 (0.004) 2,9294,341,721 h 2 (same sex) 0.471 (0.005)1,7951,187,837 h2 (male) 0.465 (0.005)2,0951,732,622 h2 (female) 0.472 (0.005)1,9571,539,582 c 2 0.174 (0.004) 2,7714,272,318 c 2 (same sex) 0.189 (0.005)1,7691,185,116 c 2 (male) 0.157 (0.004)1,9881,519,148 c 2 (female) 0.169 (0.005)1,9251,516,192 2(r MZ − r DZ) 0.593 (0.008) 9,5685,169,879 2(r MZ − r DZ) (same sex) 0.581 (0.008) 9,5684,316,578 2(r MZ − r DZ) (male) 0.593 (0.010) 4,5182,110,200 2(r MZ − r DZ) (female) 0.569 (0.010) 4,3602,240,403 2r DZ − r MZ 0.042 (0.007) 9,5685,169,879 2r DZ − r MZ (same sex) 0.055 (0.006) 9,5684,316,578 2r DZ − r MZ (male) 0.025 (0.008) 4,5182,110,200 2r DZ − r MZ (female) 0.057 (0.008) 4,3602,240,403 r, correlation; MZ, monozygotic twins; DZ, dizygotic twins; MZM, monozygo\ tic twins, male; MZF, monzygotic twins, female; DZSS, dizygotic twins, same sex; DZM, dizygo\ tic twins, both male; DZF, dizygotic twins, both female; DOS, dizygotic twins, opposite sex; h 2, heritability; c 2, proportion of variance due to shared environmental variation; estimate, estimate based on random-effects meta-analysis; n traits, number of investigated traits; n pairs, number of dependent twin pairs. The pairs are not independent, as the same or an overlapping sample of twins may have been\ used for multiple traits and across multiple studies.

npg © 2015 Nature America, Inc. All rights reserved. 7 0 6 V O L U M E 4 7 | N U M B E R 7 | J U LY 2 0 1 5 Nat u r e G e Ne t i c s A nAl y s i s we found that monozygotic twin correlations were larger than dizy - gotic twin correlations. Using the weighted estimates of r MZ and r DZ across all traits, we showed that, on average, 2 r DZ − r MZ = 0.042 (s.e.m. = 0.007) ( Table 1), which is very close to a twofold differ - ence in the correlation of monozygotic twins relative to dizygotic twins ( Supplementar y Figs. 11 and 12). The proportion of single Blood pressure funct. Conduct dis. Depr.

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funct. Weight maint.funct. rMZ 0.47 0.34 0.430.520.85 0.60 0.40 0.50 0.47 0.790.36 0.59 rMZM 0.630.29 0.58 0.42 0.45 0.60 0.88 0.48 0.50 0.34 0.59 rMZF 0.410.440.52 0.46 0.90 0.52 0.29 0.50 0.710.720.35 0.63 rDZ 0.260.23 0.270.17 0.31 0.490.16 0.23 rDZSS 0.140.09 0.290.250.49 0.25 0.19 0.33 0.33 0.14 0.33 rDZM 0.220.08 0.46 0.21 0.13 0.38 0.50 0.28 0.23 0.10 0.24 rDZF 0.100.310.19 0.22 0.49 0.23 0.12 0.32 0.430.380.15 0.35 rDOS 0.210.24 0.330.15 0.25 0.18 0.12 0.14 0 0.2 0.4 0.6 0.8 1.0 Age 65+ years Blood pressure funct. Conduct dis. Depr.

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funct. Weight maint.funct. rMZ 0.60 0.63 0.50 0.52 0.49 0.71 0.78 0.55 0.94 0.69 0.67 0.70 0.77 0.82 0.55 0.71 0.85 0.75 0.51 0.87 rMZM 0.62 0.56 0.49 0.59 0.69 0.78 0.58 0.94 0.60 0.66 0.75 0.80 0.510.83 0.86 0.52 0.85 rMZF 0.55 0.63 0.520.55 0.72 0.75 0.53 0.91 0.55 0.69 0.75 0.84 0.500.83 0.86 0.50 0.87 rDZ 0.31 0.38 0.31 0.31 0.32 0.45 0.30 0.49 0.38 0.23 0.41 0.56 0.59 0.35 0.29 0.44 0.38 0.21 0.42 rDZSS 0.36 0.43 0.36 0.42 0.17 0.36 0.45 0.36 0.58 0.38 0.21 0.63 0.62 0.320.39 0.42 0.24 0.51 rDZM 0.44 0.36 0.33 0.27 0.29 0.43 0.35 0.50 0.36 0.21 0.62 0.64 0.310.35 0.57 0.26 0.49 rDZF 0.29 0.42 0.360.42 0.41 0.46 0.36 0.54 0.35 0.31 0.58 0.60 0.330.42 0.43 0.29 0.50 rDOS 0.22 0.33 0.28 0.54 0.30 0.41 0.29 0.44 0.27 0.15 0.46 0.49 0.270.39 0.31 0.19 0.30 0 0.2 0.4 0.6 0.8 1.0 Age 12–17 years Blood pressure funct. Conduct dis. Depr.

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gland funct. Food Funct. of brain General metab. funct. Heart funct. Height High-L.

cognitive funct. Hyperkin- etic dis. Imm.

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funct. Weight maint.funct. rMZ 0.59 0.67 0.39 0.53 0.42 0.65 0.65 0.52 0.92 0.68 0.58 0.56 0.55 0.69 0.41 0.41 0.68 0.89 0.42 0.76 rMZM 0.54 0.55 0.40 0.52 0.42 0.70 0.63 0.53 0.91 0.57 0.58 0.62 0.53 0.64 0.36 0.39 0.79 0.42 0.70 rMZF 0.53 0.52 0.44 0.58 0.41 0.60 0.66 0.54 0.89 0.44 0.62 0.52 0.49 0.72 0.39 0.40 0.710.42 0.73 rDZ 0.29 0.34 0.18 0.37 0.20 0.19 0.36 0.24 0.47 0.28 0.26 0.30 0.30 0.41 0.17 0.22 0.33 0.52 0.21 0.34 rDZSS 0.30 0.43 0.21 0.34 0.24 0.24 0.33 0.29 0.53 0.27 0.31 0.38 0.36 0.44 0.16 –0.09 0.330.17 0.39 rDZM 0.25 0.36 0.18 0.32 0.22 0.39 0.32 0.21 0.53 0.34 0.35 0.29 0.31 0.37 0.14 0.25 0.330.17 0.36 rDZF 0.33 0.34 0.22 0.37 0.22 0.35 0.36 0.30 0.51 0.25 0.34 0.30 0.31 0.47 0.16 0.20 0.300.19 0.36 rDOS 0.28 0.29 0.14 0.26 0.14 0.10 0.30 0.16 0.450.250.22 0.26 0.15 0.14 0.32 0.13 0.25 –0.20 0.2 0.4 0.6 0.8 1.0 Age 18–64 years Blood pressure funct. Conduct dis. Depr.

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gland funct. Food Funct. of brain General metab. funct. Heart funct. Height Higher- level cognitive funct. Hyperkin- etic dis. Imm.

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funct. Weight maint.funct. rMZ 0.54 0.68 0.60 0.61 0.81 0.69 0.81 0.49 0.88 0.77 0.65 0.64 0.67 0.62 0.68 0.75 0.84 0.61 0.84 rMZM 0.68 0.62 0.73 0.77 0.71 0.78 0.41 0.91 0.71 0.640.580.690.62 0.83 rMZF 0.70 0.66 0.58 0.76 0.72 0.76 0.44 0.88 0.72 0.61 0.620.710.61 0.85 rDZ 0.39 0.43 0.42 0.44 0.52 0.30 0.47 0.26 0.59 0.54 0.27 0.38 0.41 0.32 0.33 0.40 0.34 0.54 rDZSS 0.41 0.44 0.39 0.22 0.55 0.33 0.52 0.33 0.61 0.58 0.24 0.41 0.43 0.410.30 0.51 0.33 0.56 rDZM 0.45 0.39 0.08 0.56 0.32 0.36 0.26 0.56 0.48 0.18 0.420.250.39 0.54 rDZF 0.42 0.40 0.32 0.62 0.32 0.54 0.26 0.59 0.50 0.20 0.360.410.37 0.56 rDOS 0.42 0.43 0.300.26 0.55 0.47 0.23 0.400.30 0.48 0.39 0.51 0 0.2 0.4 0.6 0.8 1.0 Age 0–11 years Figure 3 Twin correlations for the top 20 most investigated specific traits by age and sex. Alc., alcohol; dis., disorders; depr., depressive; endocr., endocrine; imm., immunological; funct., functions; maint., maintenance; metab., metabolic; ment. beh., mental and behavioral; spec. personal., specific personality; temp. pers., temperament and personality; tob., tobacco; r, correlation; MZ, monozygotic twins; DZ, dizygotic twins; M, males; F, females; SS, same-sex pairs only; DOS, dizygotic opposite-sex pairs. Inclusion for the top 20 most investigated traits was conditional on the reporting of r MZ and r DZ. Empty cells denote insufficient information available to calculate weighted estimates; error bars, standard errors. We note that estimates and graphs for all specific traits are available from the online MaTCH webtool.

npg © 2015 Nature America, Inc. All rights reserved. Nat u r e G eNe t i c s V O L U M E 4 7 | N U M B E R 7 | J U LY 2 0 1 5 7 0 7 A nAl y s i s Table 2 Weighted means of twin correlations and proportion ( π0) of studies that are consistent with a model where trait resemblance is solely due to additive genetic variation for the main trait domains and the top 20 most investigated traits π0 rMZ rDZ n traits Estimate Estimate (s.e.m.) n traitsn pairsEstimate (s.e.m.) n traitsn pairs All traits 5,1850.690.636 (0.002) 9,5682,563,628 0.339 (0.003) 5,2202,606,252 General trait domains Activities 620.35 0.570 (0.019) 11858,227 0.340 (0.022) 6355,864 Cardiovascular 2670.95 0.564 (0.008) 38041,669 0.295 (0.010) 26825,544 Cell 540.59 0.722 (0.022) 723,1880.523 (0.043) 541,667 Cognitive 4500.57 0.646 (0.007) 931288,867 0.371 (0.010) 454304,720 Dermatological 740.45 0.729 (0.025) 10919,509 0.402 (0.017) 7523,245 Ear, nose, throat 1650.97 0.760 (0.013) 20027,882 0.332 (0.015) 17214,222 Endocrine 1080.69 0.555 (0.017) 16210,112 0.387 (0.022) 1109,140 Environment 1450.50 0.551 (0.014) 295120,606 0.396 (0.017) 14599,137 Gastrointestinal 320.59 0.551 (0.024) 6410,982 0.274 (0.028) 3928,431 Hematological 190.65 0.764 (0.023) 505,5410.560 (0.032) 193,218 Immunological 2300.67 0.608 (0.012) 28018,051 0.357 (0.013) 23136,075 Metabolic 464 0.60 0.746 (0.005) 912210,189 0.405 (0.008) 464197,921 Neurological 7021.00 0.685 (0.005) 1,751129,076 0.289 (0.006) 70589,103 Nutritional 1100.72 0.479 (0.016) 20575,751 0.220 (0.015) 11079,188 Ophthalmological 1060.87 0.730 (0.017) 19926,139 0.385 (0.017) 10616,189 Psychiatric 1,7780.620.552 (0.004) 2,8651,232,382 0.306 (0.005) 1,7811,374,817 Reproduction 160.44 0.767 (0.034) 3412,130 0.333 (0.063) 1627,879 Respiratory 1250.74 0.697 (0.018) 18434,443 0.325 (0.019) 12751,150 Skeletal 1900.51 0.830 (0.008) 395111,282 0.504 (0.012) 191113,080 Social interactions 240.63 0.338 (0.017) 14643,501 0.267 (0.041) 2422,764 Social values 450.69 0.489 (0.030) 12052,492 0.414 (0.062) 4528,071 Top 20 investigated traits for r MZ and r DZ Blood pressure functions 1100.93 0.581 (0.010) 17920,621 0.307 (0.013) 11011,620 Conduct disorder 2160.41 0.663 (0.009) 289147,974 0.408 (0.010) 216192,651 Depressive episode 1150.60 0.454 (0.014) 17398,315 0.253 (0.015) 115121,936 Endocrine gland functions 920.72 0.538 (0.017) 1398,5330.382 (0.025) 927,295 Food 1100.72 0.479 (0.016) 20575,751 0.220 (0.015) 11079,188 Functions of brain 5940.99 0.676 (0.006) 1,01069,7220.287 (0.006) 59458,951 General metabolic functions 2190.69 0.682 (0.007) 46262,108 0.371 (0.010) 21958,338 Heart functions 1401.00 0.529 (0.009) 17415,070 0.268 (0.011) 14011,109 Height 870.29 0.908 (0.005) 12853,076 0.543 (0.008) 8768,358 Higher-level cognitive functions 1880.44 0.710 (0.009) 419152,197 0.441 (0.016) 188158,626 Hyperkinetic disorders 1000.37 0.651 (0.013) 14486,450 0.260 (0.016) 100121,139 Immunological system functions 2230.67 0.606 (0.012) 27616,703 0.357 (0.013) 22332,964 Mental and behavioral disorders due to the use of alcohol 100 0.36 0.630 (0.015) 15894,477 0.409 (0.020) 10194,196 Mental and behavioral disorders due to the use of tobacco 70 0.47 0.719 (0.016) 11051,102 0.468 (0.022) 7234,186 Other anxiety disorders 1450.29 0.548 (0.013) 191105,902 0.327 (0.016) 145153,730 Specific personality disorders 1400.93 0.448 (0.009) 16241,460 0.225 (0.007) 14033,681 Structure of the eyeball 860.91 0.735 (0.022) 12119,276 0.370 (0.019) 8613,580 Structure of mouth 1170.89 0.819 (0.010) 1277,7690.399 (0.012) 1198,493 Temperament and personality functions 5680.84 0.470 (0.008) 1,134334,190 0.234 (0.010) 568296,114 Weight maintenance functions 2150.48 0.810 (0.005) 391141,152 0.437 (0.010) 215134,867 General trait domain categories with <10 entries for π0 were excluded. For definitions of abbreviations, see Table 1. Inclusion of the top 20 investigated traits was conditional on the reporting of r MZ and r DZ. studies in which the pattern of twin correlations was consistent with the null hypothesis that 2 r DZ = r MZ was 69%. This observed pattern of twin correlations is consistent with a simple and parsimoni - ous underlying model of the absence of environmental effects shared by twin pairs and the presence of genetic effects that are entirely due to additive genetic variation ( Table 2). This remarkable fitting of the data with a simple mode of family resemblance is inconsistent with the hypothesis that a substantial part of variation in human traits is due to shared environmental variation or to substantial non-additive genetic variation. Most specific traits follow an additive genetic model Although across all traits 69% of studies showed a pattern of monozy - gotic and dizygotic twin correlations consistent with an r MZ that was exactly twice the r DZ, this finding is not necessarily representative of the majority of studies in functional domains or for every specific trait (that is, at the ICD-10 or ICF subchapter level). We thus calculated the proportion of studies consistent with 2 r DZ = r MZ within functional domains and for each specific trait and found that traits consistent with this hypothesis tended to cluster in specific functional domains ( Supplementar y Tables 27 –29 ). A pattern of twin correlations npg © 2015 Nature America, Inc. All rights reserved. 7 0 8 V O L U M E 4 7 | N U M B E R 7 | J U LY 2 0 1 5 Nat u r e G e Ne t i c s A nAl y s i s consistent with 2 rDZ = r MZ was most prominent for traits included in the neurological, ear, nose and throat, cardiovascular and oph - thalmological domains, with 99.5%, 97%, 95% and 87% of studies, respectively, being consistent with a model where all resemblance was entirely due to additive genetic variance. In only 3 of 28 general trait domains were most studies inconsistent with this model. These domains were activities (35%), reproduction (44%) and dermatologi - cal (45%) ( Table 2 and Supplementary Table 27 ). Of the 59 specific traits (ICD-10 or ICF subchapter classifications) for which we had sufficient information to calculate the proportion of studies consistent with 2 r DZ = r MZ , 21 traits showed a proportion less than 0.50, whereas for the remaining 38 traits the majority of individual studies were con - sistent with 2 r DZ = r MZ (Supplementary Table 29 ). Of the top 20 most investigated specific traits, we found that for 12 traits the majority of individual studies were consistent with a model where variance was solely due to additive genetic variance and non-shared environ - mental variance, whereas the pattern of monozygotic and dizygotic twin correlations was inconsistent with this model for 8 traits, sug - gesting that, apart from additive genetic influences and non-shared environmental influences, either or both non-additive genetic influ- ences and shared environmental influences are needed to explain the observed pattern of twin correlations ( Table 2). These eight traits were conduct disorders, height, higher-level cognitive functions, hyper - kinetic disorders, mental and behavioral disorders due to the use of alcohol, mental and behavioral disorders due to the use of tobacco , other anxiety disorders and weight maintenance functions . For all eight traits, meta-analyses on reported variance components resulted in a weighted estimate of reported shared environmental influences that was statistically different from zero ( Supplementary Table 21).

Comparison of weighted twin correlations for these specific traits resulted in positive estimates of 2 r DZ − r MZ , except for hyperkinetic disorders, where 2 r DZ − r MZ was −0.130 (s.e.m. = 0.034) on the basis of 144 individual reports and 207,589 twin pairs, which suggests the influence of non-additive genetic variation for this trait or any other source of variation that leads to a disproportionate similarity among monozygotic twin pairs.

DISCUSSION We have conducted a meta-analysis of virtually all twin studies pub - lished in the past 50 years, on a wide range of traits and reporting on more than 14 million twin pairs across 39 different countries.

Our results provide compelling evidence that all human traits are heritable: not one trait had a weighted heritability estimate of zero.

The relative influences of genes and environment are not randomly distributed across all traits but cluster in functional domains. In general, we showed that reported estimates of variance components from model-fitting can underestimate the true trait heritability, when compared with heritability based on twin correlations. Roughly two-thirds of traits show a pattern of monozygotic and dizygotic twin correlations that is consistent with a simple model whereby trait resemblance is solely due to additive genetic variation. This implies that, for the majority of complex traits, causal genetic variants can be detected using a simple additive genetic model. Approximately one-third of traits did not follow the simple pattern of a twofold ratio of monozygotic to dizygotic correlations. For these traits, a simple additive genetic model does not sufficiently describe the population variance. An incorrect assumption about narrow-sense heritability (the proportion of total phenotypic variation due to addi - tive genetic variation) can lead to a mismatch between the results from gene-finding studies and previous expectations. If the pattern of twin correlations is consistent with a substantial contribution from shared environmental factors, as we find for conduct disorders, religion and spirituality, and education, then gene-mapping studies may yield dis - appointing results. If the cause of departure from a simple additive genetic model is the existence of non-additive genetic variation, as is, for example, suggested by the average twin correlations for recur - rent depressive disorder, hyperkinetic disorders and atopic dermatitis, then it may be tempting to fit non-additive models in gene-mapping studies (for example, GWAS or sequencing studies). However, the statistical power of such scans is extremely low owing to the many non-additive models that can be fitted (for example, within-locus dominance versus between-locus additive-by-additive effects) and the penalty incurred by multiple testing. Our current results signal traits for which an additive model cannot be assumed. For most of these traits, dizygotic twin correlations are higher than half the monozygotic twin correlations, suggesting that shared environmental effects are causing the deviation from a simple additive genetic model. Yet, data from twin pairs only do not provide sufficient information to resolve the actual causes of deviation from a simple additive genetic model.

More detailed studies may identify the likely causes of such deviation and may as such uncover epidemiological or biological factors that drive family resemblance. To make stronger inferences about the causes underlying resemblance between relatives for traits that deviate from the additive genetic model, additional data are required, for example, from large population samples with extensive phenotypic and DNA sequence information, detailed measures of environmental exposures and larger pedigrees including non-twin relationships. We note that our inference is based on twin studies published between 1958 and 2012 and that it generally applies to complex traits but does not necessarily generalize to mendelian subtypes of traits.

Most mendelian traits are rare in the population and are therefore not studied by researchers of twins because they cannot ascertain enough affected twin pairs to reliably estimate genetic parameters. In the rare case where sufficient numbers of affected twin pairs were available, the mendelian subtypes were analyzed together with the subtypes of the same trait that were due to common causes, as it was unknown whether the studied trait was a mendelian subtype. If the traits we studied were in fact a mix of mendelian and complex subtypes, our inference would be biased away from our main result because mendelian diseases tend to be dominant or recessive, not additive. In addition, there may be heterogeneity in measurement errors between studies for the same trait and between traits. A test-retest correlation would quantify measurement error when contrasted with a correla - tion between monozygotic twins, but few twin studies report such correlations in the same papers that estimate heritability. Our results provide the most comprehensive empirical overview of the relative contributions of genes and environment to all human traits that have been studied in twins thus far, which can guide and serve as a reference for future gene-mapping efforts.

URLs. ICF classification, http://apps.who.int/classifications/ ic fbrowser /; ICD-10 classification, http://apps.who.int/classifica tions/ icd10/browse/2 010/en. The data used for this manuscript have been integrated in a web application, where user-specified selections of traits can be made to apply the analyses presented in this work. The web application is called MaTCH (Meta-analysis of Twin Correlations and Heritability) and is accessible via http://match.ctglab.nl/; Gephi, http://ge phi.github.io /.

M ETHODS Methods and any associated references are available in the online version of the pap er.

npg © 2015 Nature America, Inc. All rights reserved. Nat u r e G eNe t i c s V O L U M E 4 7 | N U M B E R 7 | J U LY 2 0 1 5 7 0 9 A nAl y s i s Note: Any Supplementary Information and Source Data files are available in the on li n e v ersio n o f t h e p ap er.

A C know Le D gM en TSWe would like to thank M. Frantsen, M.P. Roeling, R. Lee and D.M. DeCristo for their contribution to collecting the full texts of selected twin studies and data entry.

This work was funded by the Netherlands Organization for Scientific Research (NWO VICI 453-14-005, NWO Complexity 645-000-003), by the Australian Research Council (DP130102666) and by the Australian National Health and Medical Research Council (APP613601).

AUTHo R ConTRIBUTI onSD.P., B.B., P.F.S. and P.M.V. performed the analyses. D.P. conceived the study. D.P., T.J.C.P. and P.M.V. designed the study. T.J.C.P. and D.P. collected and entered the data. D.P. and P.F.S. categorized traits according to standard classifications.

A.v.B. and C.A.d.L. checked data entries, and checked and wrote statistical scripts.

A.v.B. designed and programmed the webtool. D.P., T.J.C.P. and P.M.V. wrote the manuscript. All authors discussed the results and commented on the manuscript.

COMPETING FINANCIAL INTERESTS The authors declare no competing financial interests.

Reprints and permissions information is available online at http ://w ww.n atu re .c om / r e p rin ts /in dex.h tm l.

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npg © 2015 Nature America, Inc. All rights reserved. Nat u r e G eNe t i c s doi:10.1038/ng.3285 ONLINE METHODS Identifying relevant studies. We searched PubMed for studies published between 1 January 1900 and 31 December 2012 that provided twin correla - tions, concordance rates, and a heritability estimate ( h 2) or an estimate of shared environmental influences ( c 2), using monozygotic and dizygotic twins.

The following search term was used (“English”[Language] AND (“1900/01/01”[Date - Publication]: “2012/12/31” [Date - Publication]) AND twin AND “journal article”[Publication Type] AND “humans”[Filter] AND (heritability[Title/Abstract] OR “genetic influence”[Title/Abstract] OR “environmental influence”[Title/Abstract] OR “genetic factors”[Title/Abstract] OR “environmental factors”[Title/Abstract]) AND “journal article”[Publication Type]) NOT review[Title] NOT review [Publication Type]) The search was run on 31 January 2013 and again on 29 April 2013, which yielded an additional 44 publications, with the difference likely due to keywords or tags that had been added to publications in the intermittent period. The last PubMed search yielded 4,388 unique studies. From these, we deleted studies that were not relevant for the current purpose using the fol - lowing exclusion criteria: (i) studies with only monozygotic twins available; (ii) studies with no heritability estimates, twin correlations or concordances available; (iii) review studies; (iv) meta-analyses; and (v) multivariate stud - ies that provided information on completely overlapping traits and samples with previously published univariate studies. Some studies investigating h 2 for the brain (for example, voxel-based brain measures) were not included for practical purposes. These studies typically presented their results in graphs with color-coded point estimates of heritability mapped onto the brain. Such estimates could not be quantified, and these studies were thus not included. From the remaining 2,748 studies, we were able to retrieve the full text from all but 5 papers (99.8%). Of the studies without full-text availability, we included relevant information based on the abstract. An overview of authors and journals and a full reference list of all 2,748 studies are provided in Supplementary Tables 30 –32 .

Primar y information obtained from each study. From every study, we retrieved basic information on the PubMed ID, the authors, the trait as named in the study and the year of publication. In addition, the following informa - tion was retrieved: • Country of origin of the study population. We used standard ISO country names, and where possible data entry was done separately for each country investigated in the study.

• Age group of the study population. The study population was classified into four age cohorts on the basis of the average age of the included sample: age >0 and <12; age ≥12 and <18; age ≥18 and <65; and age ≥65. • Monozygotic and dizygotic twin correlations. Twin correlations were entered as provided in the study and could be calculated as intraclass, Pearson, polychoric or tetrachoric correlations or on the basis of least- squares or maximum-likelihood estimates. When available, we entered the twin correlations separately for males and females (i.e., monozygotic male (MZM), monozygotic female (MZF), dizygotic male (DZM), dizygotic female (DZF) and dizygotic opposite-sex (DOS) pairs). If correlations were not available for males and females separately, we entered the MZ and DZ correlations, i.e., the correlations based on both sexes. In cases where it was clear that the dizygotic correlation was based on same-sex twins only, we entered the dizygotic same-sex (DZSS) correlation.

• Estimates of heritability ( h2) and shared environmental component ( c 2), under the full ACE (or ADE) model. We entered ‘ h 2_FULL’ and ‘c 2_FULL’, on the basis of estimates under the full ACE (including additive genetic and shared and non-shared environmental influences) or ADE (including additive and non-additive genetic and non-shared environmental influ - ences) model. When an ACE model was fitted, the estimate for A was entered in ‘ h 2_FULL’ and the estimate for C was entered in ‘ c 2_FULL’.

When an ADE model was fitted, the estimates of A and D were summed and entered for ‘ h 2_FULL’ and zero was entered for ‘ c 2_FULL’. When esti- mates were provided separately for males and females, they were entered separately. In the case of multivariate analyses, univariate estimates were always preferred to allow comparison across studies. • Estimates of heritability ( h2) and shared environmental component ( c 2), under the best-fitting ACE (or ADE) model. We entered ‘ h 2_BEST’ and ‘ c 2_BEST’, on the basis of estimates under the best-fitting ACE or ADE model as provided in the study. When an ACE model was the best-fitting model, the estimate for A was entered in ‘ h 2_BEST’ and the estimate for C was entered in ‘ c 2_ BEST’. When an ADE model was the best-fitting model, the estimates of A and D were summed and entered for ‘h 2_ BEST’ and zero was entered for ‘ c 2_ BEST’. When estimates were provided separately for males and females, they were entered separately. In the case of multivariate analyses, univariate estimates were always preferred to allow comparison across studies. In cases where estimates for the best-fitting model were not directly provided but information available in the paper indicated that the best-fitting model was AE (or CE or E), we entered zero for ‘ c 2_ BEST’ and missing for ‘ h 2_ BEST’ (when the best-fitting model was an AE model), missing for ‘ c2_ BEST’ and zero for ‘ h 2_ BEST’ (when the best-fitting model was a CE model), and zero for ‘ c2_ BEST’ and zero for ‘ h 2_ BEST’ (when the best-fitting model was described to be an E model). • The total number of twin pairs as used for each entered estimate.

• Whether the study was a classical twin study. All 2,748 studies in the database included monozygotic and dizygotic twins. However, a classi - cal twin study was defined as a study that involved only reared-together monozygotic and dizygotic twins. From studies that included siblings, extended families, adoptees or reared-apart twins, only estimates based on the reared-together twin sample were used for the meta-analyses. Most of the non-classical twin studies did provide twin correlations for the classical twin design and were thus included in the meta-analysis for twin correla - tions. When A and C estimates were based on extended twin designs, they were excluded from the meta-analyses.

• The method used for estimating the variance components. We entered the statistical method used for estimating the variance components, which included, for example, ANOVA, Bayesian, maximum-likelihood (ML), DeFries-Fulker regression, least-squares (LS) or intrapair differences. We also listed a dichotomized version of this indicating whether the method used was ‘ML or LS’ or ‘not ML or LS’, for all other methods. In the meta-analyses for h 2 and c 2 estimates, only those based on maxi - mum likelihood or least squares were included. • Whether the trait was dichotomous or continuous. Traits measured as 0 or 1, as well as traits measured on a quantitative scale but dichot - omized before analysis, were listed as dichotomous. All other traits, including ordinal traits, were listed as ‘not dichotomous’ and treated as continuous.

• Whether the study involved ascertainment for the trait. When the trait under investigation was the same trait that was used to select probands, the study was listed as ‘ascertained’.

• Number of concordant and discordant pairs. In cases of dichotomous traits, the total numbers of pairs for discordant and concordant affected pairs were entered separately for each zygosity. In cases of dichotomous traits that were not ascertained, the number of concordant unaffected pairs was also entered.

• Prevalence. In cases of dichotomous traits, the population prevalence, sep - arately for monozygotic and dizygotic twins when available, was entered.

Prevalence was based on what was provided in the study or was calculated using (2 c + d)/2 n, where c is the number of concordant affected pairs, d is the number of discordant pairs and n is the total number of pairs in non-ascertained traits.

Thus, provided that there was availability, the statistics in Supplementary Table 1 were obtained for each trait reported on in every study. When the five basic twin correlations were available ( r MZM , rMZF , rDZM , rDZF and r DOS ), we calculated r MZ, rDZSS and r DZ, using the weighted average via Fisher z transformation and using sample size as weights. In situations where r MZM (or r MZF ) was exactly 1 (or −1), we subtracted (or added in the case of −1) 0.00001 to the correlation to ensure non-problematic Fisher z transformation.

Sample sizes of MZ, DZSS and DZ were obtained by summing the sample npg © 2015 Nature America, Inc. All rights reserved. Nat u r e G eNe t i c s doi:10.1038/ng.3285 sizes of MZM and MZF, of DZM and DZF, and of DZM, DZF and DOS, respectively. Estimates of h 2 and c 2 were calculated across sex as the n -weighted average across the separate male and female estimates, when available. For the number of concordant and discordant pairs, MZ, DZ and DZSS were calculated on the basis of the numbers available for MZM, MZF, DZM, DZF and DOS. Prevalences for pooled entries were calculated as an n-weighted average.

Data entry checks. Studies were entered and cross-checked for obvious typos by T.J.C.P. and D.P. After initial data entry and initial cross-checking, all data points were manually checked (D.P.) by looking up the entered values in the original paper. In addition, automatic checks were run (D.P. and B.B.) to identify typos, strange outliers or obvious mistakes. These checks included: (i) identifying highly unlikely values (clear typos, for example, correlation of 120); (ii) testing whether the sum of h 2 and c 2 was <100; (iii) testing for strange discrepancies between estimates from the full and best-fitting models; and (iv) checking for outliers on the basis of extreme sample size and extreme χ2 values for rejecting the null hypothesis that either 2 × ( r MZ − r DZ) or 2 × r DZ − r MZ is equal to zero.

Classification of traits. After data entry, all traits were manually classified using the ICF. The ICF is the framework of the World Health Organization (WHO) for health and disability and provides the conceptual basis for the definition, measurement and policy formulations for health and disability. It is a universal classification of disability and health for use in health and health-related sectors. ICF belongs to the WHO family of international classifications, the best-known member of which is the ICD-10. ICD-10 provides an etiological framework for the classification of diseases, disorders and other health conditions, whereas ICF classifies functioning and disability associated with health conditions. The ICD-10 and ICF are therefore comple - mentary (see URLs). Most traits investigated in twin studies concern healthy functioning (for example, cognitive function, social attitudes, body height and personality) and were classified according to ICF. In cases where the studied traits were diseases or symptoms of disease, ICD-10 was used. Traits were given two hierarchical classifications corresponding to the ICF or ICD-10 hierarchical structure, using the chapter structure (for example, b1) and the level directly under the chapter (for example, b110), which corresponds to the code for the actual disease (ICD-10) or trait (ICF). Six new classes at the chapter level and 17 new classes at the subchapter level were created to accommodate traits that could not be classified under either ICF or ICD-10. For the chapter level, the created classes were cell, func - tion of DNA, functions of the nervous system, medication effects, mortality and structure of DNA. For the subchapter level, the classes created were all-cause mortality, cell cycle, cell growth, diazepam effects, expression, function of brain, gene expression, height, methylation, mortality from heart disease, mtDNA, physical appearance, receptor binding, sister chromatid exchange, structure of DNA, telomeres and X inactivation. In addition to the two standard ICF or ICD-10 classification levels, we added a general classification of functional trait domains. We thus clas - sified all traits using a 3-level scheme that included 28 broad, functional domains, 54 ICF or ICD-10 chapter-level classes and 313 subchapter-level classes. A small proportion of studied traits (<0.1%) could not be classi - fied meaningfully on the chapter level (two traits) or the subchapter level (three traits). There were 326 unique combinations across the 3 levels of trait categorization ( Supplementar y Table 33 ). All analyses were con- ducted on all entries of each of the three levels of classification. In addition, we analyzed all traits together. Although this is unspecific in terms of diseases or traits, it provides a general over view of the relationship between monozygotic and dizygotic twin correlations and shows general patterns of, for example, sex and age differences. The most specific level was the subchapter level, which was the actual ICD-10 diagnosis or a similar ICF classification for normal functioning, reflecting specific traits such as cleft lip, hyperkinetic disorders or higher-level cognitive function. As researchers do not necessarily adhere to the ICD-10 or ICF trait nomen - clature, traits with the same subchapter classification could have different trait names in the original study : for example, for higher-level cognitive function, the original studies included the trait names total IQ score, cognitive ability, intelligence or ‘g’.

Tests for publication bias. Publication bias can occur when studies that report relatively large heritability estimates or high twin correlations are more likely to be submitted and/or accepted for publication than studies that report more modest effects. Such a publication bias would lead to an overestimation of the true twin correlations or the true heritability and environmental estimates.

We used several standard statistical tools to aid in identifying and quantify - ing possible publication bias, including inspection of funnel plots, Begg and Mazumdar’s test 2 0, Egger’s regression test 21 and Rosenthal’s fail-safe N 22.

Meta-analysis methods of twin correlations and variance components. We used the DerSimonian-Laird (DSL) random-effects meta-analytical approach with correlation coefficients as effect sizes, as described by Schulze 2 3 and implemented in the R package metacor. This function transforms a correlation to its Fisher z value with corresponding standard error before the meta-analysis. This method is preferred over conducting a meta-analysis directly on the correlations because the standard error of a twin correlation is a function of not only sample size but also the correlation itself, with larger correlations having a smaller standard error. This can cause problems in a meta-analysis, as it would lead to the larger correlations appearing more precise and being assigned more weight in the analysis, irrespective of sample size. To avoid this problem, the DSL method transforms correlations to the Fisher’s z metric, whose standard error is determined solely by sample size. All n-weighted computations were thus performed using Fisher’s z metric, and the results were converted back to correlations for interpretation. The random-effects approach allows for heterogeneity of the true twin cor - relations across different studies. That is, rather than assuming that there is one true level of twin correlation, the random-effects model allows a distribution of true correlations. The combined effect of the random-effects model represents the mean of the population of true correlations. For computational reasons, correlations of −1 and 1 were converted to −0.99999 and 0.99999 before meta- analysis. We set a threshold of at least five pairs of twins available per estimate and at least two studies available per category. Meta-analyses were conducted for each category of all three levels of classification. We note that twin samples used in different publications were not independ - ent. For example, studies using Australian twins are predominantly based on twins from the Australian Twin Registry. These studies sometimes include dif - ferent subsamples but may also include completely overlapping samples used to investigate different traits. As participants are anonymous, it is not possible to determine the extent of overlap in the studies included in our analyses. We thus assumed independency of samples in the meta-analyses. This assump - tion leads to an underestimation of the variance of weighted estimates and an overestimation of their precision. We expect that the dependency of study samples is lowest at the specific level of the ICD-10 or ICF subchapters and highest for the general functional domains.

Meta-analysis for dichotomous, non-ascertained traits. In the DSL random-effects model, the standard error of a correlation is calculated on the basis of the provided n (pairs). The estimated standard errors for continuous traits are correct, but for dichotomous traits the resulting standard error is incorrect. That is because twin correlations for non-ascertained, dichotomous traits are typically based on three categories of pairs: concordant, unaffected (CON −), discordant (DIS) and concordant, affected (CON +) pairs. Whereas the total number of participating pairs is the sum of these, the information that determines the twin correlation and its significance is mostly derived from the latter two categories. For non-ascertained, dichotomous traits, we calculated the study-specific tetrachoric twin correlation on the basis of the contingency table (i.e., CON −, DIS and CON + pairs), under the assumption that the dichotomous traits represent latent variables that follow a bivariate normal distribution 2 4. We used a maximum-likelihood estimator described by Olsson 2 5, implemented in the R function polycor, to calculate the study- specific twin correlation and its standard error. As our meta-analysis required twin correlation and sample size (not standard error) as input and we wanted to be able to pool across continuous and dichotomous traits, we calculated the ‘effective’ number of pairs on the basis of the obtained standard error. npg © 2015 Nature America, Inc. All rights reserved. Nat u r e G eNe t i c s doi:10.1038/ng.3285 The effective number of pairs was defined as the number of pairs that produced the exact same standard error within the DSL meta-analyses as the standard error obtained from the contingency table.

Meta-analysis for dichotomous, ascertained traits.

For ascertained traits, it was not possible to calculate the twin correlations and standard errors on the basis of the contingency table, as the traits included only pairs with at least one proband. Without information on the number of concordant, unaffected pairs, the prevalence of the affected status would be required to calculate a twin correlation. We used the algorithms derived from Falconer 2 6 and Smith 27.

Again, for practical purposes, calculated standard errors were transformed to an effective number of pairs for use in the DSL meta-analysis.

Proportion of studies consistent with specific hypotheses. We estimated the proportion of studies that were consistent with H 0: 2 × (r MZ − r DZ) = 0 ( π0( h )) and the proportion of observations consistent with H 0: 2 × r DZ − r MZ = 0 ( π0( c )), using the Jiang and Doerge method 2 8, as well as the q-value method 2 9.

Authorship network analysis. We used the approach more fully described previously 3 0. Briefly, we retrieved from PubMed the full Medline listing for all twin studies included in this meta-analysis using NCBI eutils. The output was parsed to capture the names of all authors. The twin study author list was manually reviewed to resolve clear inconsistencies in the spelling of the names of authors between publications. Gephi was used to construct a network to understand twin study publication patterns. For clarity, we removed individuals who had published only one paper (i.e., we required authorship on ≥2 papers). The substructure of the network was investigated by estimating community membership modules using the Louvain method 3 1 implemented in Gephi. 20. Begg, C.B. & Mazumdar, M. Operating characteristics of a rank correlation test for publication bias. Biometrics 50, 1088–1101 (1994). 21. Egger, M., Davey Smith, G., Schneider, M. & Minder, C. Bias in meta-analysis detected by a simple, graphical test. Br. Med. J. 315, 629–634 (1997). 22. Rosenthal, R. The file drawer problem and tolerance for null results. Psychol. Bull. 86, 91–106 (1979). 23. Schulze, R. Meta-Analysis: A Comparison Of Approaches (Hogrefe & Huber, 2004). 24. Drasgow, F. in Encyclopedia of Statistical Sciences (eds. Kotz, S., Read, C.B., Balakrishnan, N. & Vidakovic, B.) Vol. 7, 68–74 (John Wiley & Sons, 2006). 25. Olsson, U. Maximum likelihood estimation of the polychoric correlation coefficient. Psychometrika 44, 443–460 (1979). 26. Falconer, D.S. The inheritance of liability to certain diseases, estimated from the incidence among relatives. Ann. Hum. Genet. 29, 51–76 (1965). 27. Smith, C. Concordance in twins: methods and interpretation. Am. J. Hum. Genet. 26, 454–466 (1974). 28. Jiang, H. & Doerge, R.W. Estimating the proportion of true null hypotheses for multiple comparisons. Cancer Inform. 6, 25–32 (2008). 29. Storey, J.D. & Tibshirani, R. Statistical significance for genomewide studies. Proc. Natl. Acad. Sci. USA 100, 9440–9445 (2003). 30. Bulik-Sullivan, B.K. & Sullivan, P.F. The authorship network of genome-wide association studies. Nat. Genet. 44, 113 (2012). 31. Blondel, V., Guillaume, J., Lambiotte, R. & Lefebvre, E. Fast unfolding of community hierarchies in large networks. J. Stat. Mech. 10, P10008 (2008).

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