Sex Differences in Left-Handedness: A Meta-Analysis of 144 Studies
Department of Experimental Psychology, University of Oxford, Oxford, United Kingdom.Psychological Bulletin (Impact Factor: 14.76). 10/2008; 134(5):677-99. DOI: 10.1037/a0012814
Human handedness, a marker for language lateralization in the brain, continues to attract great research interest. A widely reported but not universal finding is a greater male tendency toward left-handedness. Here the authors present a meta-analysis of k = 144 studies, totaling N = 1,787,629 participants, the results of which demonstrate that the sex difference is both significant and robust. The overall best estimate for the male to female odds ratio was 1.23 (95% confidence interval = 1.19, 1.27). The widespread observation of this sex difference is consistent with it being related to innate characteristics of sexual differentiation, and its observed magnitude places an important constraint on current theories of handedness. In addition, the size of the sex difference was significantly moderated by the way in which handedness was assessed (by writing hand or by other means), the location of testing, and the year of publication of the study, implicating additional influences on its development.
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Sex Differences in Left-Handedness: A Meta-Analysis of 144 Studies
Marietta Papadatou-Pastou and Maryanne Martin
University of Oxford
Marcus R. Munafo`
University of Bristol
Gregory V. Jones
University of Warwick
Human handedness, a marker for language lateralization in the brain, continues to attract great research
interest. A widely reported but not universal finding is a greater male tendency toward left-handedness.
Here the authors present a meta-analysis of k ⫽ 144 studies, totaling N ⫽ 1,787,629 participants, the
results of which demonstrate that the sex difference is both significant and robust. The overall best
estimate for the male to female odds ratio was 1.23 (95% confidence interval ⫽ 1.19, 1.27). The
widespread observation of this sex difference is consistent with it being related to innate characteristics
of sexual differentiation, and its observed magnitude places an important constraint on current theories
of handedness. In addition, the size of the sex difference was significantly moderated by the way in which
handedness was assessed (by writing hand or by other means), the location of testing, and the year of
publication of the study, implicating additional influences on its development.
Keywords: handedness, sex differences, meta-analysis, laterality, hand preference
Two of the most striking characteristics of humankind comprise
its use of language (e.g., Lieberman, 1984, 1991) and its popula-
tion bias toward right-handedness (e.g., Annett, 2002; McManus,
1991). Handedness and the neurobiological substrate for language
are, moreover, intimately linked, with handedness being an index
for language lateralization in the brain. Knecht et al. (2000)
showed, using transcranial Doppler ultrasonography, that the in-
cidence of right-hemisphere language dominance increases lin-
early with the degree of left-handedness, from 4% in strong right-
handers to 15% in ambidextrous individuals and 27% in strong
left-handers. This relationship, along with the fact that language
lateralization is difficult to study in large populations, has led to
extensive study of handedness, even in the archaeological record
(e.g., Pobiner, 1999; Toth, 1985; Uhrbrock, 1973). Understanding
handedness is indeed an important step toward understanding the
origin of humans, as well as understanding the relationship be-
tween handedness and language dominance. It has even been
claimed that finding the genes for handedness will be the key that
unlocks the neurobiology of language (McManus, 1991).
A relatively consistent finding in the corpus of handedness
research is that men are more prone to left-handedness than
women (e.g., Annett, 1985; M. P. Bryden, 1977; P. J. Bryden &
Roy, 2005; L. J. Chapman & Chapman, 1987; Gilbert & Wysocki,
1992; Lansky, Feinstein, & Peterson, 1988; Perelle & Ehrman,
1994; Oldfield, 1971; Reiss & Reiss, 1997). However, the size of
this sex difference varies considerably from one study to another.
Moreover, a number of studies have failed to replicate this finding
(e.g., Cornell & McManus, 1992; Green & Young, 2001; Holtzen,
1994; Salmaso & Longoni, 1985).
The putative sex difference in handedness is, if reliable, one of
the most important constraints shaping our theoretical understand-
ing of human handedness. The approximate consistency observed
in the ratio of left-handedness to right-handedness is suggestive of
a genetic basis, and a number of accounts of this kind have been
proposed (e.g., Annett, 1985; Corballis, 1997; G. V. Jones &
Martin, 2000; Klar, 1996; McManus, 1985; McManus & Bryden,
1992). At least three different explanations for a sex difference in
handedness have been offered within this framework—the differ-
ential right-shift hypothesis of Annett (2002), the modifier-gene
theory of McManus and Bryden (1992), and the recessive model of
G. V. Jones and Martin (2000). These three explanations differ not
only in their proposed mechanisms (which are described in detail
later) but also in their degree of predictiveness with respect to the
issue of a sex difference in handedness. For the right-shift hypoth-
esis, the presence of a sex difference is not an essential but an
optional characteristic of handedness. For the modifier-gene hy-
pothesis, the presence of a sex difference is an essential charac-
teristic of handedness, but its magnitude is not constrained in
advance. For the recessive model, not only the existence of the sex
effect but also its magnitude are predicted. In order to inform the
evaluation and refinement of such models, there is thus a need not
only to examine whether divergence between the sexes in hand-
edness can be detected but also to obtain a reliable quantitative
characterization of the magnitude of this divergence.
Marietta Papadatou-Pastou and Maryanne Martin, Department of Ex-
perimental Psychology, University of Oxford, Oxford, United Kingdom;
Marcus R. Munafo`, Department of Experimental Psychology, University of
Bristol, Bristol, United Kingdom; Gregory V. Jones, Department of Psy-
chology, University of Warwick, Coventry, United Kingdom.
Marietta Papadatou-Pastou was supported by the State Scholarship
Foundation, Greece. We thank Francis Marriott for helpful comments on
the statistical analysis.
Correspondence concerning this article should be addressed to Marietta
Papadatou-Pastou, Department of Experimental Psychology, University of
Oxford, South Parks Road, Oxford OX1 3UD, United Kingdom. E-mail:
Psychological Bulletin Copyright 2008 by the American Psychological Association
2008, Vol. 134, No. 5, 677–699 0033-2909/08/$12.00 DOI: 10.1037/a0012814
In order to combine the results of a corpus of studies, a meta-
analysis of appropriate odds ratios has a number of advantages (see
Rosenthal & DiMatteo, 2001). In the present study, we considered
the ratio between male and female populations of the odds between
left- and right-handedness. Although the odds ratio has not gen-
erally been employed in theoretical accounts of handedness, it is
straightforward to transpose quantitative formulations into this
measure. As an example, the recessive model of G. V. Jones and
Martin (2000, 2001) may be considered. This was developed as an
extension of the proposals by McManus (1985) and Corballis
(1997, 2001) of a handedness gene with two alleles, one (chance)
associated with a chance of either handedness, the other (dextral)
associated with right-handedness. If left-handedness is assumed to
be associated recessively with genetic variation on the X chromo-
some, it would follow that there is greater potential for its occur-
rence within the male (XY) than the female (XX) population. On the
basis of previous work, the recessive model can be shown to
provide an estimated value for the odds ratio of 1.70.
does this predicted value agree with the results of existing studies?
To address this issue, and to encourage the formulation and testing
of alternative quantitative models, it is important to obtain by
meta-analysis a best estimate of the empirical odds ratio of male to
Being able to reach a definite conclusion on whether there is a
sex difference in handedness, and estimating the true size of this
effect, is also of great importance with respect to experimental
study design. A number of studies have shown that handedness
influences cognitive functions (Annett, 1992; De Sperati & Stuc-
chi, 1997; Gentilucci, Daprati, & Gangitano, 1998; Gordon &
Kravetz, 1991; M. Martin & Jones, 1998, 1999a, 1999b; McKelvie
& Aikins, 1993; Searleman, Herrmann, & Coventry, 1984;
Townsend, Carrithers, & Bever, 2001; Gunstad, Spitznagel, Luys-
ter, Cohen, & Paul, 2007). It has even been claimed that there are
cognitive weaknesses in extreme dextrals (Annett, 1993, 1999a).
Therefore, if men are found to be more likely to be left-handed
compared to women, and at the same time there is evidence that
handedness can influence cognitive processes, then it follows that
studies that compare cognitive function between the sexes may
need to control for the incidence of handedness. Indeed, Crow,
Crow, Done, and Leask (1998) found not only a modest decrement
in a number of areas of cognitive ability with extreme levels of
relative hand skill but also a more substantial decrement close to
the point of equal hand skill. Handedness also affects language
lateralization (Khedr, Hamed, Said, & Basahi, 2002; Kimura &
Harshman, 1984; Knecht et al., 2000). Again, handedness needs to
be controlled in studies comparing the sexes in language lateral-
ization. There is also evidence, although not without conflicting
findings, that there are differences in the levels of testosterone
between right- and left-handers (Moffat & Hampson, 1996; U
Tan, 1991). Therefore, the delineation of any sex difference in
handedness is likely to have implications also for psychoendocri-
Potential Sources of Sex Differences in Handedness
A number of processes can be identified that in principle may be
responsible for the observation of sex differences in handedness, in
addition to genetic mechanisms of the type already mentioned. If
such differences are multiply determined, then systematic variation
may occur among studies in the magnitude of any sex difference,
and thus in the present study we also examined whether it was
possible to dissect out specific factors that underlie such variance.
Commencing with general methodological issues, the occur-
rence of an ascertainment bias in reporting should be considered.
If null results are less likely to be reported, then it is possible that
reports of sex differences in handedness may reflect Type I errors,
with nonsignificant results more likely to remain unpublished
(S. M. Williams, 1991). On the other hand, the difference may be
real, and at least some cases where no difference was detected may
be due to a deficiency in statistical power for small sample sizes
(Porac & Coren, 1981). The failure of some studies to find a sex
difference may be due to nonrandom sampling (Lansky et al.,
1988). The sex difference may further be a measurement artifact
arising from different reactions by the two sexes to the wording of
a hand preference inventory. M. P. Bryden (1977), for example,
has reported that men tend to avoid giving extreme responses
within handedness questionnaires.
More generally, the possible role of conformity to social pres-
sures to switch writing hand from the left to right may be consid-
ered, because women have been reported to be more successful
than men in shifting their handedness (Lansky et al., 1988; Porac,
Coren, & Searleman, 1986; Shimizu & Endo, 1983; Suar, Mandal,
Misra, & Suman, 2007; A. L. Thompson & Marsh, 1976). It may
also be that in some societies there is greater pressure upon women
than men to conform to cultural norms relating to the use of the left
hand (Harris, 1990; Suar et al., 2007). There is considerable
evidence of consistent variation in values among different nations
(e.g., Johnson, Kulesa, Cho, & Shavitt, 2005; Merritt, 2000),
including variation in the area of what Hofstede (1983, 2001) has
termed the dimension of masculinity–femininity, with masculinity
and femininity characterized by, respectively, high and low levels
of differentiation between social sex roles. A complementary ap-
proach is to consider the possibility of consistency in the variation
of constitutional influence on any sex difference in handedness,
and for this type of influence it has been proposed (e.g., Risch,
Burchard, Ziv, & Tang, 2002) that the appropriate unit of analysis
is the continent rather than the nation.
Considering in more detail the potential influence of cultural
factors upon the behavior of individuals, Kirkman, Lowe, and
Gibson (2006, p. 285) have noted that “perhaps the most influen-
tial of cultural classifications is that of Geert Hofstede.” On the
The odds ratio derived from Table 1 of G. V. Jones and Martin (2001,
p. 812) is
关共1 ⫺ d)a]⫻[1⫺(1⫺d)
is the predicted odds ratio, d is the probability of a dextral
allele, and a is the probability of phenotypic expression of left-handedness.
Simplifying Equation 1,
⫽ F(d,a)/[F(d,a) ⫺ d], (2)
where the function F(d,a) ⫽ 1⫺ (1 ⫺ d)
a. On the basis of several types
of data (twin and parental inheritance, as well as sex effect), G. V. Jones
and Martin (2000) obtained estimates for d and a of 0.38 and 0.21,
respectively. Substituting these into Equation 2, the value obtained for
PAPADATOU-PASTOU ET AL.
basis of extensive surveys, Hofstede (1983, 2001) has character-
ized national cultures in terms of not only masculinity–femininity
but also individualism (social mobility, reliance on oneself), power
distance (the unequal distribution of power, social inequality), and
uncertainty avoidance (the promotion of conformity, the need for
formalization). Of these dimensions, individualism (vs. collectiv-
ism) has been investigated the most extensively and, for example,
Oyserman, Coon, and Kemmelmeier (2002) were able to meta-
analyze the results of 83 relevant studies, providing qualified
support for the approach. On the basis of a detailed consideration
of measurement issues, Schimmack, Oishi, and Diener (2005) have
argued that the empirical data in fact clearly demonstrate the
reliability and validity of individualism as an index of cultural
differences. Nevertheless, extensive probing of individualism and
collectivism continues with, for example, the proposal that it is
important to distinguish between relational collectivism and group
collectivism (Brewer & Chen, 2007).
In the case of the cultural dimension of masculinity and femi-
ninity, one important application is the hypothesis of Arrindell et
al. (2004) that the tendency toward more restricted social roles for
both women and men (i.e., greater differentiation) that character-
izes masculine cultures may tend to be associated with psycholog-
ical ill health. Arrindell et al. (2004) empirically observed a
positive relation between cultural masculinity and national levels
of agoraphobia and other fears, and similarly Arrindell, Steptoe,
and Wardle (2003) have reported the existence of a substantial
positive relation with national levels of state depression, as as-
sessed by the short-scale Beck Depression Inventory (Beck &
Beck, 1972). More generally, there is evidence that higher cultural
masculinity is associated across nations with more extreme re-
sponse style (Johnson et al., 2005). In the present context, it may
be hypothesized that if cultural forces do exert an effect upon the
development of a sex difference in handedness, then the tendency
is likely to be more pronounced in countries with greater differ-
entiation in social roles (i.e., higher cultural masculinity).
A further possibility is that a sex difference in handedness may
arise if a condition happens to be both differentially associated
with handedness and differentially distributed between the sexes.
Conditions that may be more frequent among men than women
and that may be associated with an increased incidence of left-
handedness include homosexuality (Laumann, Gagnon, Michael,
& Michaels, 1994), dyslexia (Rutter et al., 2004), and autism
(Gualtieri & Hicks, 1985). If this is so, a sex difference might not
be detected among “unaffected” samples (Lalumiere, Blanchard,
& Zucker, 2000). Alternatively, as noted previously, a sex differ-
ence in handedness may directly reflect a constitutional bifurcation
between the sexes, such as their different testosterone levels in
utero (Galaburda, Corsiglia, Rosen, & Sherman, 1987), their dif-
ferent rates of physical maturation (Maehara et al., 1988), or their
different genetic endowments (Annett, 1996, 1998; Crow, 2002;
G. V. Jones & Martin, 2000; McManus & Bryden, 1992).
A final possibility is that a sex difference in handedness may be
related to a differential exposure to birth stress. It has been pro-
posed by a number of theorists (e.g., Bakan, 1971; Bakan, Dibb, &
Reed, 1973; Coren, 1995b) that left-handedness may be viewed as
a departure from right-handedness as a consequence of complica-
tions in pregnancy and birth, and thus a logical possibility is that
male offspring are more susceptible than female offspring to such
influences. A problem with such an explanation, however, is that
a number of studies have failed to demonstrate a major relation
overall between handedness and birth stress (e.g., Bailey & Mc-
Keever, 2004; McManus, 1981). Thus Bailey and McKeever
(2004) found that, of a wide range of potential stressors, only
maternal age was related significantly to left-handedness, and even
maternal age could account for a proportion of the total incidence
of left-handedness of the order of only 1%.
Sex Differences and the Assessment of Handedness
There are a number of other factors that, though known to
influence the observed incidence of handedness, have not been
previously studied in relation to sex differences but that may,
nonetheless, be of importance. The way in which handedness is
assessed is among the most prominent of these factors (Bishop,
1990). Assessment methods can be largely grouped into hand
preference inventories and hand proficiency measures; the former
assess which hand is preferred over the other for a number of
everyday activities (e.g., Oldfield, 1971), whereas the latter mea-
sure the relative proficiency of the two hands in performing skilled
activities. These two types of assessment are correlated, though
only imperfectly so (Bishop, 1989). Writing hand has frequently
been used as criterion, and it has been noted by Perelle and Ehrman
(1994) that self-classification of handedness also usually devolves
to writing hand. But even when restricted to hand preference
inventories, which is the method most widely used both in exper-
imental and clinical settings, it is still the case that different
instruments produce reliable differences in patterns of distribution
(Provins, Milner, & Kerr, 1982). The proportion of individuals
allocated to each handedness group may further vary depending
upon the questionnaire length, the content of the questionnaire, and
the nature of the response permitted to each item (M. P. Bryden,
1977; Gureje, 1988; Holder, 1992; Peters, 1992). In addition, the
incidence of handedness depends upon the choice of handedness
categories used: These may either be discrete, usually right and
left, with writing hand as the criterion (e.g., McManus, 1984), or
continuous, in which case the partition of the continuum into two
or more parts depends on an arbitrarily adopted criterion (e.g.,
Annett, 1994; Hardyck & Petrinovich, 1977; Maehara et al., 1988).
Until now, it has been unclear whether the manifestation of a sex
difference in handedness is sensitive to the presence of one or
more of these assessment characteristics.
Similarly, there is evidence that factors pertaining to the char-
acteristics of the populations under study may account for system-
atic variation in handedness, but the possible impact upon sex
differences has been relatively neglected. A number of special
populations (e.g., people with conditions including epilepsy, au-
tism, Down’s syndrome, and schizophrenia) have been reported to
have increased percentages of left-handedness relative to the nor-
mal population (Bishop, 1990; Boucher, 1977; Colby & Parkinson,
1977; Geschwind & Behan, 1982; Geschwind & Galaburda, 1987;
Lewin, Kohen, & Mathew, 1993; Lishman & McMeekan, 1976;
London, 1990; Merrell, 1957; Nasrallah, Keelor, & McCalley-
Winters, 1983; Sommer, Ramsey, Kahn, Aleman, & Bouma,
2001). The age of participants has also been shown to influence the
overall distribution of handedness.
Age appears to modulate the incidence of left-handedness in the
direction of an approximately linear decrease with age in both
sexes (Annett, 1973; Ashton, 1982; Brackenridge, 1981; Coren &
SEX DIFFERENCES IN LEFT-HANDEDNESS
Halpern, 1991; Dargent-Pare´, De Agostini, Mesbah, & Dellatolas,
1992; Dellatolas et al., 1991; Gilbert & Wysocki, 1992; Lee-
Feldstein & Harburg, 1982; Maehara et al., 1988; McGee &
Cozad, 1980; Porac & Coren, 1981; Salmaso & Longoni, 1985;
Schachter, Ransil, & Geschwind, 1987; Smart, Jeffery, & Rich-
ards, 1980). This effect has been attributed, amongst other factors,
to the underreporting of left-handedness in parents by their off-
spring (Kang & Harris, 1996). Similarly, a higher incidence of
left-handedness has been observed in college student samples
when compared with general population samples (Annett, 1973;
Briggs & Nebes, 1975; Peterson, 1979; Saunders & Campbell,
1985; Spiegler & Yeni-Komshian, 1983). Moreover, differences in
handedness due to location have been demonstrated (Porac, Rees,
& Buller, 1990; Raymond & Pontier, 2004), even though these
differences may only reflect cultural pressures against the use of
the left hand, resulting in low incidences of reported left-
handedness in strict or conforming societies (Lansky et al., 1988;
A. L. Thompson & Marsh, 1976). Variation in social pressures
may also be responsible for subtle secular changes in handedness
distributions. For example, McManus and Bryden (1992) divided
studies into three groups on the basis of their year of publication,
and found that the incidence of left-handed offspring among the
children of two left-handed parents was almost twice as high in the
earliest group as it was in the latest group.
Scope of Studies
It is apparent from the foregoing that a wide range of work bears
on the issues of whether there is a reliable difference between the
sexes in the distribution of handedness and, if so, what is the
overall magnitude of the difference, and what systematic influ-
ences upon it, if any, can be detected. In terms of the general area,
McManus (1986) estimated that about 5,000 articles had been
published on lateralization by 1985; updating the survey, McMa-
nus (1991) reported that between 1960 and 1989, 6,564 articles
were cited in Psychological Abstracts under the headings of cere-
bral dominance, handedness, and lateral dominance, with 1,047
being cited under the heading of handedness alone. We entered the
same search terms in PsycINFO, and we found that for the period
between 1989 and 2006, 8,757 articles were cited under the search
term of (cerebral dominance) OR handedness OR (lateral domi-
nance), and 2,195 were cited when handedness alone was used as
a subject heading. This high rate of growth in the literature makes
it increasingly difficult to have a clear picture of the research field.
A conventional literature review could not, therefore, hope to
handle such an abundance of data. A meta-analysis, on the other
hand, allows the results of a large collection of studies to be
analyzed statistically in an integrated manner (Glass, 1976). The
major advantage of meta-analysis is that it provides a discipline for
summing up research findings; the actual data of the research
studies are collected, coded, and interpreted using statistical meth-
ods similar to those used in primary data analysis, allowing for the
explanation of inconsistencies as well as the discovery of moder-
ators (Rosenthal & DiMatteo, 2001). Moreover, it allows for
ascertainment bias to be detected, which can exist when false
results are produced by nonrandom sampling. At the same time,
meta-analysis can preserve all the valuable aspects of narrative
reviews (Egger & Smith, 1997; Rosenthal & DiMatteo, 2001).
The present study is a meta-analysis of studies that have
assessed the incidence of handedness in men and women. The
first goal was to provide a definitive test of the hypothesis that
there is a sex difference in the incidence of handedness and, in
particular, to estimate the overall magnitude of any such dif-
ference. The second goal was to assess whether systematic
variation occurs in the size of the sex difference observed
among different studies and, in particular, to what such varia-
tion may be attributed. Candidate factors examined include the
assessment instrument used, the number of questionnaire items,
the type of response categories allowed, the year of publication
of the study, whether the study’s main purpose was to measure
handedness, whether the data were collected by self-report, and
the location and educational status of the participants. The
meta-analysis in addition directly assesses whether ascertain-
ment bias has played a role in the reporting of a sex difference
We located the studies that were entered into the meta-analysis
using the following procedure. We searched the computerized
reference database PubMed MEDLINE at PUBMED (NLM) using
the search terms (handedness AND (sex OR gender)) NOT (animal
OR child OR adolescent OR infant OR imaging OR functional OR
structural) via the EndNote (Version 8) citation management
software package (Thomson ISI ResearchSoft, 2004), and we
searched the online database PsycINFO using the terms handed-
ness OR hand. We scanned the cited literature of all articles that
were eligible for inclusion, and as more articles were obtained, we
searched their references for pertinent articles as well. In addition,
we hand-searched the bibliographies of six important books in the
area of handedness in order to ensure that no major studies had
been overlooked (Annett, 1985; Beaton, 1985; Corballis, 1983;
Herron, 1980; McManus, 2002; Porac & Coren, 1981). Data
collection ended in September 2007.
Inclusion in the meta-analysis of sets of data specifying hand-
edness and sex was subject to the following constraints.
1. Studies were excluded if their participants had been se-
lected or were motivated to take part on the basis of their
handedness (e.g., Casey, Pezaris, & Nuttall, 1992; Lake
& Bryden, 1976; Liederman & Healy, 1986; L. E. Tan,
1983), which typically occurred in order to increase the
proportion of left-handed participants. Members of
groups that had been reported to be specially selected
(twin persons, persons with homosexual orientation, and
persons with pathological conditions) were excluded, but
data from participants acting as their controls were in-
cluded (e.g., Cannon et al., 1995).
2. Participants were required to be over the age of 16 years.
Even though handedness is usually considered to be
established around 3–7 years of age (Hardyck, Goldman,
& Petrinovich, 1975; McManus et al., 1998; Raymond &
Pontier, 2004), this limitation was considered necessary
PAPADATOU-PASTOU ET AL.
in order to avoid possible developmental effects. An
exception was made for a few studies that had grouped
data across ages (such as 15–70) where the majority of
participants were over 16 (e.g., Ellis, Ellis, & Marshall,
3. A number of studies had included two or more samples
from different geographical areas or from different age
groups; in such cases the data sets were treated as
separate. Where the sample was subdivided into cate-
gories not meaningful for this meta-analysis (e.g., his-
tory of birth stress), then it was considered as a single
4. Data were required to have been broken down by sex in
a comprehensive way (i.e., in the text or in tables). In
some cases the data reported could not be used, for
example, in studies in which handedness for each sex was
reported only as laterality quotients (e.g., Merckelbach,
de Ruiter, & Olff, 1989; Preti, Sardu, & Piga, 2007),
when data were presented in an unsuitable way (e.g.,
Falek, 1959; B. Jones, 1980; Payne, 1987), or when only
graphical representation was available, making the accu-
rate extraction of data impossible (e.g., Provins et al.,
1982; Suar et al., 2007). Thus studies that had reported
only the result of statistical testing, without stating the
actual incidence of handedness in the two sexes (e.g.,
Salmaso & Longoni, 1983) were excluded.
5. Handedness was required to have been measured in terms
of preference, not performance. Studies were included,
however, if hand preference had been assessed by asking
participants to perform a simple action in order to ob-
serve the preferred hand.
6. Several studies duplicated data already published (e.g.,
Annett, 1999b; Blanchard & Lippa, 2007; Coren, 1995a).
In those cases, careful effort was made to include each
data set only once.
7. Reports had to be written in English. The study by
Aze´mar & Stein (1994) was the only exception, made
because the data were extracted from Raymond, Pontier,
Dufour, and Moller (1996).
Participants had been classified by the original authors pri-
marily in one of three different ways—as right- or left-handed
(R–L); as right-, mixed-, or left-handed (R–M–L); or as right-
or non-right-handed (R–nonR). In a few studies more com-
plex classifications were used (i.e., right-, right mixed-, left
mixed-, or left-handed; strong right-, mixed-, or strong left-
handed; strong right-, moderate right-, mixed-, moderate left-,
or strong left-handed; as well as a seven-class classification).
In the first case the results were converted to R–L, whereas in
the latter cases they were converted to R–M–L. In the case of
Salmaso & Longoni (1985), who provided 10 laterality
classes, a cutoff point was introduced in order to classify the
participants as R–L.
The categorical and interval variables whose possible effects
were examined included the following.
Instrument. Handedness has been assessed in a number of
different ways, which may be grouped as follows. First, identifi-
cation of writing hand has been widely employed. Second, there
are a number of questionnaires and inventories, which can them-
selves be grouped into four types. Third, there is self-classification,
for example, “Do you consider yourself to be left-, (mixed-), or
right-handed?” Fourth, a specific action (other than writing) is
observed, for example, holding a racket (including information
from official sports records).
Considering further the second category (questionnaires and
inventories), these may be further grouped as follows: (a) the
Edinburgh Handedness Inventory (10-item version only) of Old-
field (1971), (b) the four handedness items from the Lateral Pref-
erence Inventory of Porac and Coren (1981), (c) the Hand Prefer-
ence Questionnaire (8-item, 10-item, 12-item, and 23-item
versions) of Annett (1970), and (d) the modification of Annett’s
Hand Preference Questionnaire by Briggs and Nebes (1975). Ex-
amination reveals considerable overlap in content among these
questionnaires. Briggs and Nebes employed the same items as did
Annett, with a wider response scale (5-point instead of 2 or 3).
Further, the Edinburgh and 12-item Annett instruments have half
or more of their items in common (hand preference for writing,
throwing, and using scissors, a toothbrush, a broom, and a match),
and similarly the majority of Porac and Coren items overlap with
either the Edinburgh or Annett scales or both. It may also be noted
that the questionnaires and inventories differ from the other in-
struments in a potentially important manner because they focus on
not only unimanual but also bimanual activities (e.g., which is the
upper hand when using a broom).
Finally, in those cases where percentages on a number of items
were given, without reporting a laterality quotient, the information
on writing hand was used (e.g., McFarland & Anderson, 1980;
Merrell, 1957). For the Aggleton and Wood (1990) study, the
information on the hand throwing a 10-pin bowling ball was used
for both groups (professional bowling players and a control group),
as information on writing hand was reportedly collected for the
control group but not included in the article.
Location. As noted earlier, extensive use has previously been
made of the four cultural dimensions of masculinity, individual-
ism, power distance, and uncertainty avoidance, introduced by
Hofstede (1983, 2001). Estimated values on all four dimensions
were reported by Hofstede (1983) for 50 countries; in the case of
masculinity, for example, Scandinavian (or Nordic) countries ex-
hibited the least differentiation, with Sweden and Norway occu-
pying the 1st and 2nd ranks, respectively, and Finland (often
grouped with these countries), the 7th rank. A composite measure
of cultural formality may be derived, following Medland, Perelle,
De Monte, and Ehrman (2004), by subtracting individualism from
the sum of masculinity, uncertainty avoidance, and power distance.
The potentially moderating effects of all five measures were in-
vestigated in the present analysis.
A number of human population genetic studies have claimed
that genetic differentiation is greatest when defined on a continen-
tal basis (see Risch et al., 2002, for a review) and have suggested
the categorization of populations into five major groups: Cauca-
SEX DIFFERENCES IN LEFT-HANDEDNESS
sians, Africans, East Asians, Pacific Islanders, and Native Amer-
icans. Here we used only the three former groupings, as none of
the studies reported the inclusion of participants in the latter two
groups. The Caucasian group included participants from Europe,
North America, West Asia, and Australia. Rarely was information
about ancestry reported, and instead it was replaced with a degree
of approximation by the location in which the study took place. Of
course, variation among countries may also reflect cultural differ-
Educational status. Educational status was grouped at two
levels, the higher comprising individuals who had entered college
(college students, faculty members, and professionals). Mixed
samples (i.e., samples including both students and their family
members) and samples for which no demographic information was
given were not included within the analysis of the moderating
effect of educational status.
Year of publication. For investigating possible moderation in
terms of secular change, the year of publication was entered
numerically for each study.
Type of response categories. Four groupings were employed:
(a) R–L responses (including the cases where participants had to
tick a box next to a picture showing hand posture), (b) 3-point
scale responses, including variations of right– both–left, right–
equally–left, right–ambidextrous–left and always right–either
right or left–always left, (c) 5-point scale responses, and (d) the
response scheme used in the original version of the Edinburgh
Handedness Inventory. This scheme features two columns labeled
right and left. Participants are asked to indicate their preferences in
the use of hands in the activities listed in the questionnaire by
marking ⫹ in the appropriate column. Where preference is so
strong that participants would never try to use the other hand
unless absolutely forced to, they are asked to put ⫹⫹. If they are
indifferent, they are asked to put ⫹ in both columns.
Other variables. Additionally, information on whether the
measurement of handedness was the main purpose of the study and
whether the data were collected by self-report were also extracted
from the studies, using a yes/no coding, and the number of ques-
tionnaire items used was entered numerically.
Not all the studies reported information for each of the above
moderator variables. The mean age of the participants was reported
in fewer than 25% of the data sets, insufficient for mean age to be
utilized as a moderator variable.
We analyzed the data using the Comprehensive Meta-Analysis
(Version 2; Borenstein, Hedges, Higgins, & Rothstein, 2005)
Five separate analyses were carried out, based on five dif-
ferent conceptions of left-handedness. The first three of these,
here termed left-handedness (extreme), left-handedness (forced
choice), and mixed handedness, adopted progressively more lax
criteria for left-handedness; the fourth used the criterion of
non-right-handedness; and the fifth, termed left-handedness
(total), was an inclusive analysis. The five specifications were
1. Left-handedness (extreme): Extreme left-handers repre-
sent the participants who were classified as left-handers
in data sets where an R–M–L classification was em-
2. Left-handedness (forced choice): Left-handers by forced
choice represent the participants who were classified as
left-handers in data sets where an R–L classification was
3. Mixed-handedness: Mixed-handers represent the partici-
pants who were classified as mixed-handers in studies
where an R–M–L classification was employed. In prin-
ciple, an alternative possibility would have been to define
a group containing both mixed-handers and left-handers,
but this would have introduced nonorthogonality into any
comparison with the left-handedness (extreme) group.
Accordingly, it should be noted that for mixed-
handedness as it was in fact constituted, the relatively lax
criterion for membership of the left group (namely, being
to the left of extreme right-handedness) was combined
with the further constraint of excluding extreme left-
4. Non-right-handedness: Non-right-handers represent the
participants who were classified as non-right-handers in
data sets where an R–nonR classification was employed.
It was noted that considerable uncertainty exists over
criterion placement in studies adopting this form of clas-
sification, which may range from not at all right-handed
to not extremely right-handed (i.e., embracing approxi-
mately the full range of the three preceding membership
5. Left-handedness (total): This comparison was de-
signed to assess the overall presence of left-
handedness. Information was used from all the data
sets, regardless of whether they classified their partic-
ipants in terms of R–L, R–nonR, or R–M–L. It was
recognized that the increase in sample size (and hence
statistical power) obtained by combining over the three
classification schemes could be at least partially offset
in principle by greater heterogeneity in group mem-
bership. Thus in order to ensure that the left-handed
(total) group was defined as strictly as possible, in the
R–M–L case it was comprised only of the left-handed
members. In those cases where the same participants
were classified twice using more than one type of
measurement (e.g., both by means of writing hand and
of a handedness inventory), the information on the
writing hand was preferred, as it is the most popular
way of measuring of handedness, and therefore it
would introduce homogeneity in the meta-analysis. If
information on writing hand was not one of the op-
tions, then data on self-classification were used (e.g.,
Reiss, Reiss, & Freye, 1998). In the case of two studies
where both the R–L and the R–M–L classifications
were available for the same measures, information
from the latter classification was used (Brito, Brito,
Paumgartten, & Lins, 1989; Saunders & Campbell,
PAPADATOU-PASTOU ET AL.
We calculated male to female odds ratios and corresponding
two-tailed 95% confidence intervals for each data set indepen-
dently and then combined them using a fixed effects model to
provide a pooled odds ratio and a test for the overall effect (Z
statistic). An odds ratio value of 1.0 corresponds to the null
hypothesis of no sex difference, whereas values greater than 1.0
indicate a larger proportion of male than female left-handers.
Moreover, each comparison was tested for heterogeneity using the
homogeneity statistic Q, and the extent of inconsistency among the
data sets’ results was assessed using the I
index. The I
be interpreted as the percentage of total variation across studies
that is due to heterogeneity rather than chance (computed negative
values are set equal to zero), and Higgins, Thompson, Deeks, and
Altman (2003) have proposed that levels of 25%, 50%, and 75%
may be described as low, moderate, and high, respectively. In the
cases of significant heterogeneity between the data sets, we re-
peated the analysis using a random effects model. Whereas a fixed
effects model assumes that all the data sets included in the meta-
analysis come from a single population, a random effects model
assumes that the included data sets are drawn from a distribution
We tested the data sets for ascertainment bias using Egger’s t
statistical test and the funnel plot graphical test. The rationale of
the latter test is that if all data sets come from a single population,
then the plot of the odds ratio against the standard error should
resemble a funnel with the diameter of the funnel decreasing (i.e.,
effect size estimates becoming more accurate) as the sample size
increases. In the absence of ascertainment bias, one would expect
a symmetrical funnel plot, and asymmetry is therefore suggestive
of the possibility of ascertainment bias. Duval and Tweedie’s
(2000) trim and fill method of correcting bias was also used. This
method aims at making the funnel plot symmetrical by omitting
and/or adding hypothetical data sets to the plot where necessary. It
then provides an adjusted estimate of the effect size that may be
compared with the original empirical value.
In examining the possible moderating effects of categorical
moderator variables, the effect sizes in the different subgroups that
form the levels of each moderator were compared by means of the
Q statistic. In examining the possible effects of interval moderator
variables, meta-regressions were performed; random effects mod-
els were used as recommended by S. G. Thompson and Higgins
(2002), with evaluation again in terms of the Q statistic.
A total of 144 studies were included for analysis, comprising
208 separate data sets and totaling 1,787,629 individuals (831,537
men, 956,092 women). Table 1 presents the details of the studies
used. Analyses were conducted first on each of the five different
types of handedness categorizations previously outlined.
Left-handedness (total). This comparison included k ⫽ 199
data sets drawn from 144 studies, adding up to n ⫽ 1,787,629
individuals (n ⫽ 831,537 men, n ⫽ 956,092 women). Fixed effects
analysis gave a pooled odds ratio (OR) ⫽ 1.28, 95% confidence
interval (CI) ⫽ 1.27, 1.30; Z ⫽ 51.40, p ⬍ .001. Significant
heterogeneity was found to exist among the data sets, Q(198) ⫽
329.10, p ⬍ .001, with small-to-moderate inconsistency between
⫽ 39.84%). A random effects model was therefore
employed, which revealed a clear difference in left-handedness
between the sexes (OR ⫽ 1.23, 95% CI ⫽ 1.19, 1.27), indicating
that the odds ratio was significantly different from 1.0 (Z ⫽ 13.17,
p ⬍ .001).
Left-handedness (extreme). This comparison included k ⫽ 51
data sets drawn from 42 studies, comprising up to n ⫽ 240,346
individuals (n ⫽ 127,516 men, n ⫽ 112,830 women). Fixed effects
analysis gave a pooled OR ⫽ 1.22, 95% CI ⫽ 1.19, 1.25; Z ⫽
14.39, p ⬍ .001. Significant heterogeneity was found to exist
among the data sets, Q(50) ⫽ 78.88, p ⬍ .01, with small-to-
moderate inconsistency between studies (I
⫽ 36.61%), indicating
that one or more variables may moderate the relationship between
sex and handedness. A random effects model was therefore em-
ployed, which revealed a clear difference in left-handedness (ex-
treme) between the sexes (OR ⫽ 1.20, 95% CI ⫽ 1.11, 1.29),
indicating that the odds ratio was significantly different from 1.0
(Z ⫽ 4.62, p ⬍ .001).
Left-handedness (forced choice). This comparison included
k ⫽ 137 data sets drawn from 94 studies, comprising up to n ⫽
358,602 individuals (n ⫽ 184,003 men, n ⫽ 174,599 women).
Fixed effects analysis gave a pooled OR ⫽ 1.23, 95% CI ⫽ 1.20,
1.26; Z ⫽ 18.17, p ⬍ .001. Significant heterogeneity was found to
exist among the data sets, Q(136) ⫽ 199.12, p ⬍ .001, with small
inconsistency between studies (I
⫽ 31.70%). A random effects
model was therefore employed (OR ⫽ 1.24, 95% CI ⫽ 1.19, 1.30),
indicating that the odds ratio was significantly different from 1.0
(Z ⫽ 10.30, p ⬍ .001).
Mixed-handedness. This comparison included k ⫽ 51 data sets
drawn from 42 studies, comprising up to n ⫽ 240,346 individuals
(n ⫽ 127,516 men, n ⫽ 112,830 women). Fixed effects analysis
gave a pooled OR ⫽ 1.16, 95% CI ⫽ 1.09, 1.23; Z ⫽ 5.10, p ⬍
.001. Significant heterogeneity was found to exist among the data
sets,Q(50) ⫽ 152.80, p ⬍ .01, with moderate-to-large inconsis-
tency between studies (I
⫽ 67.28%). A random effects model was
therefore employed, which revealed a clear difference in mixed-
handedness between the sexes (OR ⫽ 1.31, 95% CI ⫽ 1.16, 1.48),
indicating that the odds ratio was significantly different from 1.0
(Z ⫽ 4.28, p ⬍ .001).
Non-right-handedness. This comparison included k ⫽ 20 data
sets drawn from 18 studies, adding up to n ⫽ 1,198,476 individuals
(n ⫽ 524,611 men, n ⫽ 673,865 women). Fixed effects analysis
gave a pooled OR ⫽ 1.31, 95% CI ⫽ 1.29, 1.32; Z ⫽ 46.23, p ⬍
.001. Significant heterogeneity was found to exist among the data
sets, Q(19) ⫽ 43.61, p ⬍ .01, with moderate inconsistency be-
tween studies (I
⫽ 56.44%). A random effects model was there
fore employed, which revealed a clear difference in non-right-
handedness between the sexes (OR ⫽ 1.22, 95% CI ⫽1.10, 1.36),
indicating that the odds ratio was significantly different from 1.0
(Z ⫽ 3.72, p ⬍ .001).
It is apparent from these results that odds ratios were numeri-
cally greater for comparisons using more lax criteria of left-
handedness. Figure 1 summarizes graphically the odds ratios with
their 95% confidence intervals for the left-handedness (extreme),
left-handedness (forced choice), and mixed-handedness compari-
sons. As noted previously, the lax criterion for left-handedness in
this last case was combined with the exclusion of extreme left-
(text continues on page 687)
SEX DIFFERENCES IN LEFT-HANDEDNESS
Summary of the Studies Included in the Meta-Analysis
Author Year Location Sample size Odds ratio Measure of handedness
Aggleton et al. 1994 UK 1,538 0.99 Writing hand
Aggleton & Wood 1990 USA/UK 707 1.57 Hand used for bowling (official records/self-report of
Annett 1973 UK 10,932 1.03 Writing hand
Annett 1979 UK 3,034 0.75 Writing hand/filial report or report by spouses on writing
hand/12-item Annett’s Hand Preference Questionnaire3.01
Annett 1985 — 1,679 1.74 Not reported (2 data sets)/hand holding the racket (tennis
Annett 2002 — 200 1.43 Hand holding the racket
Annett 2007 UK 5,612 1.31 Writing hand
Annett & Kilshaw 1982 UK 1,550 0.91 Writing hand/8-item and 12-item Annett’s Hand
Ardila & Rosselli 2001 Colombia 6,941 1.20 Self-classification
Ashton 1982 USA 3,625 1.00 Writing hand/hand usage
Aze´mar & Stein 1994 — 2,490 1.13 Hand holding sword/foil
Bakan & Putnam 1974 Canada 400 2.47 Writing hand
Barut et al. 2007 Turkey 633 1.76 Edinburgh Handedness Inventory
Beckman & Elston 1962 Sweden 981 0.92 Not reported
Betancur et al. 1990 France 205 0.98 Modified version of 10-item Edinburgh Handedness
Birkett 1981 UK 125 0.82 10-item Edinburgh Handedness Inventory
Briggs & Nebes 1975 USA 1,599 0.95 12-item Briggs-Nebes modification of Annett’s
Brito et al. 1989 Brazil 959 2.10 10-item Edinburgh Handedness Inventory
M. P. Bryden 1977 Canada 1,106 1.38 Writing hand
M. P. Bryden 1989 Canada 794 1.56 8 items from Edinburgh Handedness Inventory
P. J. Bryden & Roy 2005 Canada 153 2.21 Writing hand
Buchtel & Rueckert 1984 USA 740 1.06 Writing hand
Cannon et al. 1995 Ireland 43 0.50 10-item Edinburgh Handedness Inventory
Carriere & Raymond 2000 Cameroon 246 1.30 Observation of which hand is used to hold the machete
Casey & Brabeck 1989 USA 433 1.08 10-item Edinburgh Handedness Inventory
Chamberlain 1928 USA 4,354 1.44 Writing hand
D. Chapman & Walsh 1973 Australia 923 0.92 Throwing a ball
L. J. Chapman &
1987 USA 5,825 1.12 13-item questionnaire
Coren 1989 Canada 1,896 1.19 4 items for handedness from Porac & Coren’s Lateral
Coren 1993 Canada 3,307 1.32 4 items for handedness from Porac & Coren’s Lateral
Coren 1995b Canada 2,596 1.20 4 items for handedness from Porac & Coren’s Lateral
Coren & Porac 1979 Canada 1,758 1.10 Writing hand
Coren & Porac 1980 Canada 4,171 1.10 4 items for handedness from Porac & Coren’s Lateral
Coren et al. 1986 Canada 1,180 1.60 4 items for handedness from Porac & Coren’s Lateral
Cornell & McManus 1992 UK 266 0.53 Writing hand
Cosenza & Mingoti 1993 Brazil 16,590 1.45 10-item Edinburgh Handedness Inventory
Cosenza & Mingoti 1995 Brazil 15,389 1.25 10-item Edinburgh Handedness Inventory
Cuff 1931 USA 109 0.76 8-item questionnaire
Curt et al. 1995 France 1,609 1.00 12-item questionnaire
Dane & Erzurumluog˘lu 2003 Turkey 326 1.36 10-item Edinburgh Handedness Inventory
Dargent-Pare´ et al. 1992 Algeria 5,064 0.92 12-item questionnaire
PAPADATOU-PASTOU ET AL.
Table 1 (continued)
Author Year Location Sample size Odds ratio Measure of handedness
De Agostini et al. 1997 Ivory Coast 2,989 1.93 Parents: filial reported
1.00 Offspring: 10-item questionnaire
DeLisi et al. 2002 USA, UK, Italy 288 3.15 23-item Annett’s Hand Preference Questionnaire
Demura et al. 2006 Japan 3,557 2.13 10-item Edinburgh Handedness Inventory
Downey 1927 USA 721 1.91 5-item questionnaire
Dronamraju 1975 India 517 2.17 Hand used to hold a brush
Elalmis & Tan 2005 Turkey 22,461 1.12 Self-classification
Elias et al. 2001 Canada 541 1.73 Self-classification
Ellis et al. 1988 USA 6,577 1.14 10-item Edinburgh Handedness Inventory
Fry 1990 USA 1,087 1.25 Parents: filial report of writing hand
0.98 Offspring: 10-item Edinburgh Handedness Inventory
1990 USA 60 3.22 12-item Annett’s Hand Preference Questionnaire
Gilbert & Wysocki 1992 USA 1,177,507 1.31 Writing hand and throwing hand
Gladue & Bailey 1995 USA 149 1.36 10-item Annett’s Hand Preference Questionnaire
Go¨testam 1990 Norway 235 0.94 Writing hand/4-item questionnaire
Green & Young 2001 UK 284 0.83 6-item questionnaire
Grouios et al. 2000 Greece 2,299 1.20 12-item Briggs-Nebes modification of Annett’s
Gunstad et al. 2007 USA 643 1.02 Edinburgh Handedness Inventory
Gur & Gur 1977 USA 200 2.98 23-item Raczkowski et al.’s questionnaire
Halpern et al. 1998 USA 152,653 1.24 Writing hand
Hannay et al. 1990 USA 1,185 0.97 10-item questionnaire
Harburg et al. 1978 USA 1,496 1.66 Self-classification
Harburg et al. 1981 USA 1,153 1.20 Writing hand
Harris & Gitterman 1978 USA 356 0.94 12-item Briggs & Nebes handedness questionnaire
Harvey 1988 UK 398 1.54 10-item Edinburgh Handedness Inventory
Hatta & Kawakami 1995 Japan 1,700 1.50 10-item questionnaire
Hatta & Nakatsuka 1976 Japan 1,199 1.95 10-item questionnaire
Heim & Watts 1976 UK 890 1.84 Writing hand
Hicks & Kinsbourne 1976 USA 2,202 1.00 Filial report of writing hand
Hicks et al. 1978 USA 728 2.05 12-item Briggs & Nebes modification of Annett’s
Hicks et al. 1980 USA 580 1.01 12-item Briggs & Nebes handedness questionnaire
Holder 1992 USA 314 1.21 Self-classification
Holtzen 1994 USA 260 0.56 5 items from the Edinburgh Handedness Inventory
Holtzen 2000 — 1,685 1.05 Hand holding the racket
1987 Belgium 128 1.52 4 items for handedness from Porac & Coren’s Lateral
Hoosain 1990 Hong Kong 556 3.11 10-item questionnaire
Ida & Bryden 1996 Japan 1,275 1.95 Writing hand
Inglis & Lawson 1984 USA 1,880 1.23 3-item questionnaire: self-classification, observation of
writing hand, and hand usage
Iwasaki et al. 1995 Japan 1,755 1.75 Writing hand
Kauranen & Vanharanta 1996 Finland 200 0.18 Self-classification
Lansky et al. 1988 USA 2,083 1.49 Writing hand
1982 USA 1,153 1.20 Writing hand
Leiber & Axelrod 1981 USA 2,477 1.38 Self-classification
Lester et al. 1982 USA 2,168 1.85 Not reported
Levander & Schalling 1988 Sweden 921 0.78 Writing hand
Lippa 2003 USA 1,056 0.84 Self-classification/3-item inventory
Maehara et al. 1988 Japan 2,459 1.23 10-item Edinburgh Handedness Inventory
Marchant-Haycox et al. 1991 UK, USA 396 1.21 9-item inventory
W. L. B. Martin & Porac 2007 Brazil 1,635 1.10 Self-classification
SEX DIFFERENCES IN LEFT-HANDEDNESS
Table 1 (continued)
Author Year Location Sample size Odds ratio Measure of handedness
Mascie-Taylor 1980 UK 386 1.24 Writing hand/7-item questionnaire
Mascie-Taylor et al. 1981 UK 141 2.56 Writing hand
McFarland & Anderson 1980 Australia 181 1.38 Writing hand
McGee 1976 USA 112 3.94 7 items from Annett (1970)
McGee & Cozad 1980 USA 1,230 1.22 10-item Edinburgh Handedness Inventory
McKeever & Rich 1990 USA 3,080 1.18 Writing hand
1986 UK 2,028 0.93 Self-classification
Merrell 1957 USA 620 1.07 Writing hand
Me´sza´ros et al. 2006 Hungary 150 1.37 Structured medical questionnaire
Morley & Caffrey 1994 UK 3,814 1.06 Writing hand/self-classification
Mustanski et al. 2002 USA 382 1.05 Self-classification
Nalc¸aci et al. 2001 Turkey 310 2.32 13-item questionnaire adapted from L. J. Chapman &
Newcombe & Ratcliff 1973 UK 823 1.40 7-item questionnaire
Newcombe et al. 1975 UK 928 1.46 7-item questionnaire
Obrzut et al. 1992 USA 318 1.50 1st factor of the Waterloo Handedness Questionnaire (14
Ofte 2002 Norway 393 0.96 5-item questionnaire
Oldfield 1971 UK 1,109 1.77 10-item Edinburgh Handedness Inventory
Overby 1994 USA 963 1.08 Self-classification
Perelle & Ehrman 1983 USA 2,404 1.05 13-item questionnaire
Perelle & Ehrman 1994 32 countries 32,039 1.28 Writing hand
Peters et al. 1981 Canada 365 1.08 4-item questionnaire
Peters et al. 2006 UK 169,230 1.24 Writing hand
Plato et al. 1984 USA 705 1.75 Writing hand/10-item inventory
Porac 1993 Canada 632 0.64 6-item questionnaire
Porac et al. 1983 Canada 900 1.14 Writing hand
Porfert & Rosenfield 1978 USA 2,107 1.05 Short questionnaire
Raymond et al. 1996 France 969 1.13 Students: writing hand
1.56 Athletes: hand holding discus, javelin, shot put, or racket
Reiss & Reiss 1997 Germany 936 1.13 4 items for handedness from Porac & Coren’s Lateral
Reiss et al. 1998 Germany 1,223 1.97 Self-classification
Rife 1940 USA 3,552 1.29 10-item questionnaire
Risch & Pringle 1985 USA 7,391 1.12 Offspring: 10-item Edinburgh Inventory and self-
Parents: filial report
Rosenstein & Bigler 1987 USA 50 1.31 10-item Edinburgh Handedness Inventory
Sakano & Pickenhain 1985 Japan 1,688 1.90 5 items from the Edinburgh Handedness Inventory
Salmaso & Longoni 1985 Italy 1,694 1.06 20-item Edinburgh Handedness Inventory
Sanders et al. 1982 USA (European,
879 0.80 Hand preference questionnaire
Saunders & Campbell 1985 USA 372 1.98 10-item Edinburgh Handedness Inventory
Schachter et al. 1987 USA 1,117 1.95 10-item Edinburgh Handedness Inventory
Searleman & Fugagli 1987 USA 277 1.43 Writing hand
Searleman et al. 1979 USA 847 1.05 14-item modified Crovitz & Zener handedness index
Searleman et al. 1984 Canada 3,709 1.48 4 items for handedness from Porac & Coren’s Lateral
Segal 1984 USA 1,577 1.13 Writing hand
Shan-Ming et al. 1985 China 432 2.30 10-item preference demonstration
Sherman 1979 USA 98 3.18 14-item Crovitz & Zener questionnaire
1983 USA 218 0.79 Filial report
Shimizu & Endo 1983 Japan 4,282 1.74 13-item questionnaire
PAPADATOU-PASTOU ET AL.
handedness, to preserve orthogonality. Further, the relatively small
number of studies reporting non-right-handedness were not in-
cluded in the comparison because of the broad range of criteria
embraced by this classification, as indicated earlier.
The data sets that were included in the left-handedness (total)
comparison were tested for ascertainment bias. No ascertainment
bias was detected using Egger’s test, t(198) ⫽ 1.21, p ⫽ .23.
Similarly, a funnel plot graphical test (see Figure 2) did not
provide visual evidence of asymmetry and hence possible ascer-
tainment bias. Duval and Tweedie’s (2000) method led to two data
sets being “trimmed” and no data sets being “filled.” The resulting
adjusted odds ratio (random effects model) was 1.23 (95% CI ⫽
1.19, 1.27), an estimation identical to the odds ratio originally
calculated (OR ⫽ 1.23, 95% CI ⫽1.19, 1.27) and hence again
providing no evidence of ascertainment bias. Finally, the fail-safe
N was calculated and found to be N ⫽ 17,408. The fail-safe N is
the number of data sets with an odds ratio of one (i.e., zero effect)
Table 1 (continued)
Author Year Location Sample size Odds ratio Measure of handedness
Singh & Bryden 1994 India 729 1.64 1st factor from the 59-item version of the Waterloo
Handedness Questionnaire (10 items)
Smith 1987 UK 350 1.64 10-item Edinburgh Handedness Inventory
Spiegler & Yeni-
1983 USA 5,448 1.24 Writing hand
. Tan 1986 Turkey 266 15.73 12-item Annett’s Hand Preference Questionnaire
. Tan 1988 Turkey 1,100 0.76 Turkish adaptation of the Edinburgh Handedness
Tapley & Bryden 1985 Canada 1,511 1.21 8-item questionnaire
Teng et al. 1979 Taiwan 2,041 2.08 12-item Edinburgh Handedness Inventory
A. L. Thompson &
1976 USA 1,299 1.79 4-item questionnaire
Wolf et al. 1991 USA 2,088 1.18 Interview
Wood & Aggleton 1989 UK 752 1.03 Hand used to hold a racket
Wood & Aggleton 1991 UK 1,240 2.26 10-item Edinburgh Handedness Inventory
Unpublished data set reported in Seddon and McManus (1993), “The Incidence of Left-Handedness: A Meta-Analysis.”
(extreme) comparison (forced choice)
Male to Female Odds Ratios
Figure 1. Graphic representation of the male to female odds ratios and corresponding two-tailed 95%
confidence intervals for the left-handedness (extreme), left-handedness (forced choice), and mixed-handedness
SEX DIFFERENCES IN LEFT-HANDEDNESS
that would be needed to be added to the existing meta-analysis for
it to be no longer significant at the conventional level of p ⬍ .05,
and thus the high value of N confirms the reliability of the
The moderating effects of the previously indicated variables
were tested within the left-handedness (total) comparison. This
comparison included data from all n ⫽ 1,787,629 participants
within k ⫽ 199 independent data sets and is therefore the most
representative as well as the most powerful one. Table 2 presents
detailed statistics for each level of the categorical variables. Since
the left-handedness (total) comparison included studies that had
classified their participants as either R–L, R–M–L, or R–nonR, it
was first confirmed that these different classifications within the
comparison did not have a significant moderating effect on the sex
difference in left-handedness, Q(2) ⫽ 1.50, p ⫽ .47, before inves-
tigating the presence of other possible moderator variables. For all
the analyses, only data sets in which pertinent information was
reported were used.
Location. There was significant heterogeneity overall among
these studies, Q(180) ⫽ 298.29, p ⬍ .001, with small-to-moderate
variation between studies (I
⫽ 39.64%). It can be seen from Table
2 that the odds of being left-handed rather than right-handed were
found to be 22%, 31%, and 60% higher for male than for female
participants among the Caucasian, African, and East Asian sam-
ples, respectively; location was found to exert a significant mod-
erating effect on the size of the sex difference in handedness,
Q(2) ⫽ 11.26, p ⫽ .01. Further examination within groups re-
vealed that significant heterogeneity remained in the Caucasian
group, Q(156) ⫽ 270.08, p ⬍ .001; I
⫽ 42.24%; but not in the
East Asian group, Q(16) ⫽ 17.68, p ⫽ .34; I
⫽ 9.50%; or the
African group, Q(6) ⫽ 3.26, p ⫽ .72; I
The Caucasian group was investigated for further moderator
effects by contrasting North American samples (k ⫽ 79) with
European samples (k ⫽ 57). Within this overall set of studies there
was significant heterogeneity, Q(135) ⫽ 230.38, p ⬍ .001; I
41.40%, with a significant moderating contrast between North
America and Europe, Q(1) ⫽ 4.66, p ⬍ .05; the odds ratio was
higher for North America (OR ⫽ 1.25, 95% CI ⫽ 1.20, 1.31) than
for Europe (OR ⫽ 1.14, 95% CI ⫽ 1.07, 1.23). However, there
remained significant low to moderate levels of heterogeneity of
variance within both the North American group, Q(78) ⫽ 122.30,
p ⫽ .001; I
⫽ 36.22%; and the European group, Q(56) ⫽ 84.95,
p ⬍ .01; I
Examining the North American group, the contrast between the
United States (k ⫽ 58) and Canada (k ⫽ 21) was not significant,
Q(1) ⫽ 0.93; significant heterogeneity remained within U.S. stud-
ies, Q(57) ⫽ 105.42, p ⬍ .001; I
⫽ 45.93%, but not within
Canadian studies, Q(20) ⫽ 16.88; I
⫽ 0%. Examining the Euro
pean group, the contrast between Scandinavia (here Norway, Swe-
den, and Finland; k ⫽ 9) and the remaining group (k ⫽ 48) reached
borderline significance, Q(1) ⫽ 3.51, p ⫽ .061. For Scandinavia,
the magnitude of the odds ratio was below unity, though not
significantly so (OR ⫽ 0.88, 95% CI ⫽ 0.67, 1.17; Z ⫽ 0.88), and
there was no significant heterogeneity, Q(8) ⫽ 4.51; I
⫽ 0%. For
the remaining group, the odds ratio was significantly above unity
(OR ⫽ 1.16, 95% CI ⫽ 1.08, 1.25; Z ⫽ 4.04, p ⬍ .001), and there
was significant heterogeneity, Q(47) ⫽ 75.39, p ⬍ .01; I
37.66%. It happened to be that the majority of the Scandinavia data
sets derived from studies in which handedness categorization had
been made on the basis of writing hand or self-classification,
whereas categorization in the majority of the non-Scandinavian
data sets had been made on the basis of questionnaires. Thus to
remove any possible confounding effect of categorization instru-
ment, the European analysis was repeated with both groups re
-3 -2 -1 0 1 2 3
Log odds ratio
Figure 2. Funnel plot of standard error on log odds male to female ratio, for the total left-handedness
PAPADATOU-PASTOU ET AL.
stricted to studies using only writing hand or self-classification.
This analysis yielded results very similar to the original one, and
therefore excluded the possibility of a confound effect. The con-
trast between the reduced Scandinavian (k ⫽ 7) and non-
Scandinavian (k ⫽ 15) groups remained at borderline significance,
Q(1) ⫽ 2.75, p ⫽ .098. For Scandinavia, the magnitude of the odds
ratio remained below unity, though not significantly so (OR ⫽
0.83, 95% CI ⫽ 0.57, 1.22; Z ⫽ 0.93), and there was no significant
heterogeneity, Q(6) ⫽ 2.75; I
⫽ 0%. For non-Scandinavia, the
odds ratio remained significantly above unity (OR ⫽ 1.17, 95%
CI ⫽ 1.05, 1.20; Z ⫽ 2.78, p ⬍ .01), and there was significant
heterogeneity, Q(14) ⫽ 25.38, p ⬍ .05; I
Meta-regression over the entire range of data sets revealed a
significant increase in the sex differences odds ratio with national
masculinity, Q(1) ⫽ 3.99, p ⬍ .05; the best-fitting linear relation
between log odds ratio and masculinity was ln(OR) ⫽ 0.00307
(masculinity) ⫹ 0.0252. The odds ratio also increased significantly
with uncertainty avoidance, Q(1) ⫽ 4.44, p ⬍ .05; the best-fitting
relation with uncertainty avoidance was ln(OR) ⫽ 0.00208 (avoid-
ance) ⫹ 0.0991. In addition, the odds ratio was not significantly
influenced by power distance, Q(1) ⫽ 3.11, but did decrease
significantly with individualism, Q(1) ⫽ 5.96, p ⬍ .05; the best-
fitting relation with individualism was ln(OR) ⫽⫺0.00218 (indi-
vidualism) ⫹ 0.378. Finally, the odds ratio increased significantly
with the composite measure of formality, Q(1) ⫽ 7.78, p ⬍ .01;
the best-fitting relation was ln(OR) ⫽ 0.00108 (formality) ⫹
Instrument. The moderating effect of the instrument used to
measure handedness also approached significance, Q(6) ⫽ 12.82,
p ⫽ .08. In view of the practical importance of the issue of
selecting an instrument for measuring handedness, this variable
was examined further. It was noted that one measure, writing hand,
stood out as by far the most frequent way to measure handedness
in our data set. Writing hand (k ⫽ 54 data sets) was therefore
compared with all the other instruments (k ⫽ 77 data sets), and the
moderating effect was found to be significant, Q(1) ⫽ 8.36, p ⫽
.02; the odds ratio for writing hand was 1.16 (95% CI ⫽ 1.10,
1.23), and the odds ratio for the other measures combined was 1.24
(95% CI ⫽ 1.18, 1.30). There remained moderate levels of heter-
ogeneity within writing hand, Q(53) ⫽ 97.24, p ⬍ .001; I
45.50%, but no significant heterogeneity within the other mea-
sures, Q(80) ⫽ 88.27, p ⫽ .25; I
⫽ 9.37%. As noted earlier,
self-classification can generally be assimilated to writing hand, and
thus these categories were also combined (k ⫽ 72) and contrasted
with questionnaire studies (k ⫽ 46), yielding a significant effect,
Q(1) ⫽ 5.93, p ⫽ .015, for writing hand and self-classification
(OR ⫽ 1.17, 95% CI ⫽ 1.12, 1.22) and for the questionnaire
measures (OR ⫽ 1.28, 95% CI ⫽ 1.21, 1.35). It was also noted
earlier that writing hand (and self-classification), unlike question-
naire measures, lacks a bimanual component. Consistent with the
lower odds ratio for writing hand (i.e., OR ⫽ 1.17), a similar value
was obtained when those studies (k ⫽ 11) employing a specific
unimanual action were analyzed (OR ⫽1.16, 95% CI ⫽ 1.01,
Male to Female Left-Handedness Odds Ratios (OR) and 95% Confidence Intervals (CI) for All
the Levels of the Moderator Variables
Moderator variable No. of data sets OR 95% CI
Right-left (R-L) 134 1.24 1.19, 1.29
Right-mixed-left (R-M-L) 49 1.19 1.10, 1.28
Right-non-right (R-nonR) 16 1.29 1.16, 1.44
Writing hand 54 1.16 1.10, 1.23
Edinburgh Handedness Inventory 27 1.28 1.19, 1.38
Porac & Coren’s Lateral Preference Inventory 9 1.28 1.17, 1.42
Annett’s Hand Preference Questionnaire 4 2.28 1.07, 4.86
Briggs & Nebes 6 1.19 0.97, 1.47
Self-classification 18 1.17 1.07, 1.27
Specific action 13 1.20 1.05, 1.37
Caucasian 157 1.22 1.18, 1.26
East Asian 17 1.60 1.36, 1.87
African 7 1.31 1.02, 1.65
College 76 1.26 1.17, 1.35
Other 103 1.23 1.18, 1.27
Right-left 37 1.26 1.19, 1.33
3-point scale 68 1.24 1.16, 1.33
5-point scale 30 1.22 1.12, 1.33
⫹ or ⫹⫹ (Edinburgh scheme) 8 1.34 1.17, 1.53
Main purpose to measure handedness
Yes 107 1.22 1.17, 1.27
No 91 1.26 1.20, 1.32
Self-report 171 1.23 1.19, 1.27
Report by others 26 1.29 1.16, 1.43
SEX DIFFERENCES IN LEFT-HANDEDNESS
1.34), with no significant heterogeneity, Q(10) ⫽ 5.90, p ⫽ .82;
Publication year. Meta-regression on the year of publication
of the studies revealed a significant linear trend of the size of the
sex difference in handedness, Q(1) ⫽ 6.44, p ⫽ .01; the best-fitting
linear relation between log odds ratio and year was ln(OR) ⫽
⫺0.00199 (year) ⫹ 4.213, equivalent to a decline in estimated
odds ratio between 1927 and 2007 from 1.46 to 1.24.
Other moderators. No significant moderating effects were
found for a number of other variables, including the educational
status of the participants, Q(1) ⫽ 0.57, p ⫽ .75; the type of
response categories used, Q(3) ⫽ 2.37, p ⫽ .66; whether the
main purpose of the study was to measure handedness, Q(1) ⫽
1.02, p ⫽ .31; and whether the data were collected by self-
report, Q(1) ⫽ 1.70, p ⫽ .43. Meta-regression on the number of
questionnaire items also did not reveal any significant effect,
Q(1) ⫽ 0.23, p ⫽ .63.
The present meta-analysis includes data on 1,787,629 individ-
uals (831,537 men, 956,092 women) extracted from 144 studies,
divided into 208 separate data sets. Tests of ascertainment bias
revealed that there was no evidence that the studies in the sample
had been distorted by preferential reporting. A comprehensive
comparison was carried out on all the data sets, termed left-
handedness (total), and this provided a best estimate of 1.23 for the
ratio of male to female left- to right-handedness odds (95% CI ⫽
1.19, 1.27). Though the odds ratio has mathematical properties
which make it peculiarly suitable for combining evidence, it is
generally recognized that simple proportions may be easier to
grasp intuitively. It can be shown that their relationship in the
present study takes the form of m ⫽ f ⫻ OR/[1 ⫹ f(OR ⫺ 1)],
where m and f are the probabilities of left-handedness in male and
female populations, respectively. As an illustration, if the inci-
dence of female left-handedness were exactly 10%, the observed
best estimate of 1.23 for the odds ratio is equivalent to an incidence
of male left-handedness of 12.0%.
In addition to the result overall, a significant sex difference was
also detected in each of four other meta-analyses carried out on
smaller sets of data. There was some indication that the odds ratio
tended to increase as the criterion for left-handedness became
more lax, with estimates for male to female odds ratios ranging
from 1.20 (95% CI ⫽ 1.11, 1.29) for the left-handedness (extreme)
analysis (left-handers identified in data sets where an R–M–L
classification was employed) to 1.24 (95% CI ⫽ 1.19, 1.30) for the
left-handedness (forced choice) analysis (left-handers identified in
data sets where an R–L classification was employed), and 1.31
(95% CI ⫽ 1.16, 1.48) for the mixed-handedness analysis (mixed-
handers identified in data sets where an R–M–L classification was
These findings constrain substantially the range of acceptable
predictions by quantitative theories of human handedness. Thus it
was noted earlier that the odds ratio predicted on the basis of the
parameter values estimated for their recessive model by G. V.
Jones and Martin (2000) takes the value of 1.70. Because the
theoretical odds ratio lies outside the confidence intervals estab-
lished here, it is apparent that reconsideration of this model is
indicated. We return later to the model of G. V. Jones and Martin
and to other theories of the sex effect in handedness, but first it is
important to note the general point that the present analyses
provide information not only on the overall magnitude of the effect
but also on the occurrence of systematic variations from this.
Over the whole sample, a small-to-moderate proportion of the
variability across studies was found to be due to heterogeneity
rather than to chance. Three factors were found to moderate
significantly the size of the sex difference odds ratio, namely, the
way in which handedness had been assessed, the year of publica-
tion of the study, and the location of the participants. The sex
difference was larger when handedness was assessed using meth-
ods other than the recording of writing hand (or, equivalently,
writing hand together with self-classification), in earlier rather than
later studies, and in East Asian rather than Caucasian and African
samples. Whereas the unidirectional consistency of the sex effect
in handedness across comparisons is compatible with a genetic or
other constitutional basis, the occurrence of these systematic vari-
ations in magnitude suggests the importance also of factors within
the environmental domain. Thus although the geographical factor
may not distinguish between these two domains, the significant
variation in human behavior observed here as a function of the
study’s date (over a period of only 80 years) points directly to the
latter, environmental domain. Discouragement of left-handed writ-
ing has occurred within some educational systems, thus reducing
the apparent incidence of left-handedness when writing hand is
used as criterion and compressing the potential range of sex
difference to be observed. Consistent with this, the sex difference
was found to be significantly less for the writing hand criterion
(OR ⫽ 1.16) than for the other measures (OR ⫽ 1.24). Handedness
questionnaires and inventories, in contrast, include areas that are
relatively unlikely to be influenced by social pressure, in particular
those involving both hands. For example, sweeping with a broom,
dealing playing cards, guiding a thread through the eye of a needle,
and unscrewing the lid of a jar were all included by Annett (1970,
2002). Further analyses here were consistent with the hypothesis
that the inclusion of activities requiring bimanual coordination
within handedness questionnaires is responsible for the observa-
tion of an enhanced sex difference. The present work thus suggests
that it may in future research be informative to compare explicitly
the role of handedness in influencing the performance of uni-
manual versus bimanual activities. Archaeological evidence from
humeri bones suggests that the proportions of these two types of
activity have varied quite widely in different cultures, with sub-
stantial differences between women and men (e.g., Bridges, 1989;
Sla´dek, Berner, Sosna, & Sailer, 2007), and theoretical work has
suggested that it may indeed be left-handed advantages in predom-
inantly male fighting interactions that have driven genetic varia-
tion in handedness (Billiard, Faurie, & Raymond, 2005).
Within a contemporary context, pressures for social conformity
that may lead to apparently reduced incidences of left-handedness
are known to vary over place and time (Harris, 1990; see also
Provins, 1997). Such pressures appear to be stronger in relatively
conservative societies and in earlier periods (see, e.g., W. L. B.
Martin & Porac, 2007; Searleman & Porac, 2003) and would result
in greater sex differences if they bore more heavily on female than
on male left-handedness. The odds ratios here for the East Asian,
Caucasian, and African groups were, respectively, 1.60, 1.22, and
1.31. The first of these groups comprised Japanese and Chinese
participants, and here reputed pressure toward the use of the right
PAPADATOU-PASTOU ET AL.
hand has been noted by Medland et al. (2004). Within the second
group, it was observed that the general sex– handedness effect was
numerically, though not significantly, reversed in Scandinavia
(here represented by Sweden, Norway, and Finland). This phe-
nomenon does not seem to have been previously noted but appears
to be related to wider societal characteristics, as tabulated by
Hofstede (1983, 2001), because these countries exhibit extremely
low levels of cultural masculinity (i.e., differentiation in social sex
roles). Across all countries, the magnitude of the sex difference in
handedness was found here to increase significantly with increases
in national masculinity. In addition, the sex difference increased
significantly with increases in uncertainty avoidance and de-
creased significantly with increases in individualism; further meta-
regression indicated a significant positive dependence of sex dif-
ference upon a composite measure of cultural formality. These
findings provide support for the hypothesis that the sex difference
in handedness is at least partly determined by cultural values that
have been shown to influence a considerable range of psycholog-
ical processes across different nations (e.g., Arrindell et al., 2003;
Kirkman et al., 2006; Schimmack et al., 2005).
Much remains to be understood about possible social influences
upon handedness. Do the sexes tend to experience different social
pressures with regard to handedness, and, in particular, do they in
turn respond to these pressures differently? Further work may
show whether such differences in response can arise either on the
basis of differential tendencies toward social compliance (as
shown by, e.g., Gabriel & Gardner, 1999; Van Vugt, De Cremer,
& Janssen, 2007) or alternatively on the basis of differential
patterns of neurobiological flexibility (as shown by, e.g., Boghi et
al., 2006; Garavan, Hester, Murphy, Fassbender, & Kelly, 2006).
No significant effects upon the sex difference in handedness
were detected for a number of other potential moderating factors,
including the educational status of the participants, the number of
questionnaire items used, the type of response categories used,
whether the main purpose of the study had been to measure
handedness, and whether the data were collected by self-report.
The power of meta-analytic studies is a function of the number of
entered studies (Hunter & Schmidt, 1990) and thus was reduced in
the case that only a subset of studies had reported information
pertinent to a particular factor. Nevertheless, even the smallest of
these analyses, that for response category, included as many as 143
Only studies that assessed hand preference were included in the
present analyses; studies measuring hand skill were excluded.
Although there is a debate about the relative value of these two
manifestations of handedness (Bishop, 1989; Morgan & Corballis,
1978), preference may be considered to be primary, because the
presence of preference asymmetry in the absence of skill asym-
metry has been demonstrated in children with autism (McManus,
Murray, Doyle, & Baron-Cohen, 1992). Moreover, differences in
hand preference but not in hand skill have been detected between
individuals with schizophrenia or schizoaffective disorder and
their unaffected relatives (DeLisi, Svetina, Razi, Shields, Well-
man, & Crow, 2002). Nevertheless, preference can be reliably
correlated with measures of performance (Annett, 1976), even
though these correlations are not perfect (0.6 to 0.7; Todor &
Doane, 1977). An upper limit on the magnitude of such correla-
tions appears to be provided by at least two factors: first, the
difference in distributions between preference and skill measures
(generally negatively skewed and normal, respectively), and sec-
ond, the relatively low levels of reliability of measures of relative
hand skill (Hiscock & Chapieski, 2004). This correlation does
suggest however that the results of the present meta-analysis may
have relevance also with regard to sex differences in hand skill.
Another important distinction in handedness research is that
between direction versus degree of handedness (Dellatolas, Pas-
cale, & Curt, 1997). Direction and degree of handedness have been
given different biological and psychological interpretations (Mc-
Manus, 1983). The findings of a neuroimaging study using fMRI
suggest that these aspects are independent and that they are coded
separately in the brain (Dassonville, Zhu, Ugurbil, Kim, & Ashe,
1997). These two measures are nevertheless confounded in studies
that used statistical tests that are unable to differentiate between
them or that merely reported the mean laterality score across their
sample. It would require large-scale studies using the same hand-
edness questionnaires reporting the number of women and men
gaining each score. In the present meta-analysis, only direction of
hand preference was taken into account, as limited information on
the degree of hand preference was reported in the included studies.
The closest the current meta-analysis can approach this issue is the
intriguing finding that the sex difference in left-handedness ap-
pears to lie upon a continuum; the stricter the criterion for left-
handedness, the lower the male to female odds ratio and thus the
smaller the difference between the two sexes. This is counterin-
tuitive, as one would expect the largest sex difference to be found
using the more strict criteria for left-handedness. Evidence pre-
sented in this article suggests rather that perhaps men are in fact
more prone to ambidexterity rather than to extreme left-
handedness compared to women. A phenomenon of this kind may
be linked to differences in callosal morphology. There is some
evidence that the corpus callosum tends to be larger in left-handed
and ambidextrous people than in right-handed people and that this
is modulated by sex (e.g., Habib et al., 1991; Witelson, 1989).
However, such effects have not always been observed (e.g.,
Kertesz, Polk, Howell, & Black, 1987; Steinmetz et al., 1992), and
recent work (Luders et al., 2003) has implicated the influences of
sex and handedness upon more subtle relations between callosal
connections and sulcus asymmetry.
Showing that a sex difference in handedness is present in every
comparison representing different conceptions of left-handedness
provides support for the theories that explain handedness with
biological factors pertinent to sexual differentiation (i.e., genetic
theories, hormonal theories, and theories on the rate of somatic
maturation). We believe that further investigation on the sex dif-
ferences in handedness should be along the lines of those theories.
It should be noted that suggesting that the sex differences are best
explained by biological factors does not preclude that environmen-
tal factors could be moderating the size of male to female odds
ratios in different populations. In fact, our findings provide some
evidence that cultural pressures do moderate the size of the sex
differences in handedness.
What are the implications of the present results for the three
genetic explanations of the sex effect in handedness considered
earlier, namely, the differential right-shift hypothesis of Annett
(2002), the modifier-gene theory of McManus and Bryden (1992),
and the recessive model of G. V. Jones and Martin (2000)?
Annett’s right-shift theory (Annett, 1972, 1998) postulates a gene
for left cerebral dominance that increases the probability of right-
SEX DIFFERENCES IN LEFT-HANDEDNESS
handedness. The right-shift gene is a systematic influence on
human asymmetry that works by impairing the growth of the right
hemisphere in early life and incidentally weakening the left hand
and channeling language and speech functions to the left side. A
sex difference for handedness is not integral to the theory but can
be interpreted as due to the displacement of a chance distribution
of asymmetry farther to the right in women than in men by about
20% (Annett, 1999b). This might be because the right-shift gene is
more penetrant in women than in men (Annett, 1973, 1985) or
because women are slightly more mature at birth than men and, as
the expression of the right-shift positive (RS⫹) gene is considered
to be a function of growth in utero, this difference in maturation
accounts for the fact that women are slightly more often right-
handed (Annett, 1996, 1998) than men.
The theory of McManus (1984) proposed dextral and chance
alleles for handedness. Homozygotes for the dextral allele are all
right-handed, whereas homozygotes for the chance allele show
chance levels of handedness, with the heterozygous phenotype
intermediate. This theory has no provision for a sex effect and thus
was amplified by McManus and Bryden (1992), who proposed that
a second relevant gene, a sex-linked recessive modifier gene on the
X chromosome, can suppress the dextrality gene. As men have
only one copy of the X chromosome, they are more likely to
express the suppressor gene and to show an increased rate of
phenotypic left-handedness. Insofar as this modifier-gene hypoth-
esis makes an integral prediction of the occurrence of a sex-
difference in handedness, it receives greater support from the
present findings than does the differential right-shift theory, which
could accommodate a sex difference of any degree, including
none. As noted by Roberts and Pashler (2000), the occurrence of
observations that are consistent with a theory provides strong
support for the theory only if the theory also prohibits the occur-
rence of alternative observations.
The present findings are particularly informative for the single-
gene recessive model of G. V. Jones and Martin (2000). The
proposal of an X-linked gene, one of whose alleles is associated
with the chance of left-handedness, is supported by striking ma-
ternal and grandparent effects (G. V. Jones & Martin, 2001, 2003).
The maternal effect refers to the finding of McKeever (2000) that
the probability of a son being left-handed is influenced by the
handedness of his mother but not that of his father. The grandpar-
ent effect refers to the finding of Klar (1996) that the child of two
right-handed parents, one of whom is the child of two left-handed
grandparents, is as likely to be left-handed as is the child of a
left-handed parent and a right-handed parent. As with the modifier-
gene hypothesis, the recessive model’s integral prediction of the
existence of a sex difference receives significant support from the
present findings. But for this model, there are quantitative as well
as qualitative implications. As noted previously, the observed
magnitude of the odds ratio was lower than predicted. In principle,
this result could be accommodated by appropriate adjustment of
parameter values, in particular by assigning a lower value to d, the
probability of the dextral allele. However, the previously estimated
value of d was shown by G. V. Jones and Martin (2000) to be
quantitatively consistent with other important aspects of handed-
ness including twin concordance and parental inheritance, and it
remains to be investigated whether this would be the case for
suitably revised parameter values. An alternative possibility is the
introduction of differential male and female penetrance parame-
ters, reflecting a phenotypic influence of the kind shown to exert
significant moderating effects in the present analyses. Epigenetic
processes that confer heritable changes in a manner which in
principle may be affected by the environment (e.g., DNA methyl-
ation) are also of interest (e.g., Bird, 2007; Hong, West, & Green-
In terms of physical chromosomal mapping, Laval et al. (1998)
have also argued for an X-linked recessive influence on handed-
ness. ProtocadherinXY, a gene located in the Xq21.3/Yp region of
homology between the X and the Y chromosomes, is another
candidate gene for asymmetry (Crow, 2002; N. A. Williams,
Close, Giouzeli, & Crow, 2006), and sequence differences be-
tween the X and Y copies of protocadherinXY could account for
the gender differences in lateralization. Most conclusive support
for a genetic role in handedness is provided by the recent discovery
(Francks et al., 2007) that the Leucine-rich repeat transmembrane
neuronal 1 (LRRTM1) gene was significantly associated with
handedness in a set of dyslexic siblings.
Differences in physical maturation rate between the sexes may
also be of relevance. A delayed rate of maturation has been found
to be associated with an increased incidence of left-handedness
regardless of sex (Coren, Searleman, & Porac, 1986; Mulligan,
Stratford, Bailey, McCaughey, & Betts, 2001). Moreover, there is
evidence that male children develop a right-hand preference later
than do female children (Archer, Campbell, & Segalowitz, 1988;
Carlson & Harris, 1985; Humphrey & Humphrey, 1987). As boys
mature later than girls, the sexual component of handedness could
thus be influenced by maturation rate (Maehara et al., 1988).
Finally, possible hormonal influences on the sex difference in
handedness may be noted. According to the theory of Geschwind
and Galaburda (1985a, 1985b, 1985c, 1987), testosterone has
differential effects in the development of the cerebral hemispheres,
acting so as to increase the probability of right hemisphere dom-
inance by slowing down left hemisphere growth and thus acting as
a source of left-handedness (but see also M. P. Bryden, McManus,
& Bulman-Fleming, 1994). Taking into account that the develop-
ing male brain is exposed to higher concentrations of testosterone
than the female brain, then increased left-handedness in men is to
be expected. In our view, genetic, maturational, and hormonal
theories are not mutually exclusive. It could be the case that they
focus on different aspects of the same phenomenon, maturation
being intertwined with hormonal changes that are controlled by
genetic factors, all resulting in the different neural organization of
the two sexes, with the neural mechanisms serving praxis and
speech being less lateralized in the left hemisphere in men than in
women (Kimura & Harshman, 1984).
Overall, the present study provides a powerful test of the hy-
pothesis of a sex difference in handedness. We have shown using
objective statistical procedures that the sex difference is robust for
all the commonly used conceptions of left-handedness, with men
having significantly greater odds of being left-handed than women
(except, it appears, in Scandinavia). In addition, it has been shown
that the possible range in the magnitude of the sex difference is
sufficiently narrow to impose a significant new constraint on the
quantitative genetic modeling of handedness. The size of the sex
difference is moderated by means of the instrument used to mea-
sure handedness (writing hand or not), the location in which testing
occurred, and the year of publication of the studies. Further, the
size of the sex difference in left-handedness may lie on a contin-
PAPADATOU-PASTOU ET AL.
uum from the strictest to the most lax measures of classifying
participants as left-handers. The evidence reviewed here suggests
that the sex difference in hand preference has its basis in innate
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Received June 14, 2007
Revision received April 7, 2008
Accepted April 10, 2008 䡲
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SEX DIFFERENCES IN LEFT-HANDEDNESS
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- "In fact, some samples are skewed in terms of sex ratio; for example, Mandal et al. (1999) included only male participants. The age of the participants is another possible moderator, as age has been found to modulate the incidence of lefthandedness in the direction of an approximately linear decrease with age in both sexes (for a review see Papadatou-Pastou et al., 2008 ). In addition to the above, the sign language used by the participants of each study might exert a moderating effect. "
- "Men are 3 to 4 times more likely than women to be affected by stressful prenatal environments, leaving them susceptible to the development of psychopathologies including paraphilias (Cantor, 2012; Gualtieri & Hicks, 1985; Kraemer, 2000). Handedness offers a glimpse into prenatal brain organization and early perturbations in development, with men being more likely to be non-right-handed than women (Papadatou-Pastou, Martin, Munafò, & Jones, 2008 ), and individuals who have neurological disorders being more likely to be non-right-handed (Lewin, Kohen, & Mathew, 1993). Rahman and Symeonides (2008) found that men who scored high on paraphilic fantasies also had higher incidence of non-right-handedness compared with men who scored low. "
- "Left-handedness is more commonly found in triplets (7.1%) and twins (8.1%) than in single births, while ambidextrous triplets (6.4%) are more prevalent than ambidextrous twins (3.4%) or singletons (3.5%) (Vuoksimaa, et al., 2009; Orlebeke, et al., 1996). Other studies from the past two decades have affirmed earlier research that reported left-handedness was more prevalent in twins than in single births, and that left-handedness was also more common in males than females (Davis and Annett, 1994; Papadatou-Pastou et al., 2008). Davis and Annette in particular found their results to be strong empirical evidence for the RS handedness theory, first proposed in 1978 by Annette and then revised in 1985. "