Intelligence and handedness: Meta-analyses of studies on intellectually disabled, typically developing, and gifted individuals

Neuroscience and Biobehavioral Reviews 56 (2015) 151–165

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Intelligence and handedness: Meta-analyses of studies on intellectually disabled, typically developing, and gifted individuals Marietta Papadatou-Pastou a,b,∗,1 , Dimitra-Maria Tomprou a,1 a Research Centre for Psychophysiology and Education, School of Education, National and Kapodistrian University of Athens, 27 Deinokratous Str, 106 75 Athens, Greece b Cognition and Health Research Group, Medical Sciences Division, Department of Experimental Psychology, University of Oxford, Tinbergen Building, 9 South Parks Road, Oxford OX1 3UD, UK

a r t i c l e

i n f o

Article history: Received 23 February 2015 Received in revised form 6 June 2015 Accepted 19 June 2015 Available online 2 July 2015 Keywords: IQ Language laterality Behavioral laterality Intellectual disability Intellectual giftedness Hand preference

a b s t r a c t Understanding the relationship between cerebral laterality and intelligence is important in elucidating the neurological underpinnings of individual differences in cognitive abilities. A widely used, behavioral indicator for cerebral laterality, mainly of language, is handedness. A number of studies have compared cognitive abilities between groups of left- and right-handers, while others have investigated the handedness prevalence between groups of different cognitive abilities. The present study comprises five meta-analyses of studies that have assessed the handedness prevalence in (a) individuals with intellectual disability (ID) of unknown/idiopathic nature compared to typically developing (TD) individuals, and (b) individuals with intellectual giftedness (IG) compared to TD individuals. Nineteen data sets totaling 16,076 participants (5795 ID, 8312 TD, and 1969 IG) were included in the analyses. Elevated levels of atypical handedness were found to be robust only for the ID to TD comparison. Findings constrain the range of acceptable theories on the handedness distribution for different intelligence levels. © 2015 Elsevier Ltd. All rights reserved.

Contents 1. 2.

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Introduction . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 151 Method . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 154 2.1. Study selection . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 154 2.2. Moderators . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 154 2.3. Statistical analysis . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 155 Results . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 155 3.1. Results of the ID-TD analyses . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 157 3.2. Results of IG-TD analyses . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 160 Discussion . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 161 Conclusions . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 163

1. Introduction The relationship between intelligence and handedness has attracted research interest since the turn of the 20th century when

∗ Corresponding author at: Research Centre for Psychophysiology and Education, School of Education, National and Kapodistrian University of Athens, 27 Deinokratous Str, 106 75 Athens, Greece. E-mail addresses: [email protected] (M. Papadatou-Pastou), [email protected] (D.-M. Tomprou). 1 Both authors contributed equally. http://dx.doi.org/10.1016/j.neubiorev.2015.06.017 0149-7634/© 2015 Elsevier Ltd. All rights reserved.

increased rates of atypical handedness (left-handedness or nonright-handedness) were reported in individuals with intellectual disability (ID; Binet and Vaschide, 1897; Doll, 1916; Gordon, 1921) and is still debated today (e.g., Beratis et al., 2013; Mellet et al., 2014; Nicholls et al., 2012). The effects of individual differences, such as handedness, on intelligence are important because general intelligence test scores are associated with key life outcomes, including school achievement (e.g., Deary et al., 2007; Johnson et al., 2006), socioeconomic success (e.g., Strenze, 2007), and job performance (e.g., Gottfredson, 1997), albeit the relationship is not always linear (e.g., Zagorsky, 2007). Moreover, if populations with special

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educational needs, such as ID individuals, are found to present elevated levels of atypical handedness, this will constrain the range of acceptable theories on the handedness distribution for different intelligence levels and contribute to the understanding the condition. Handedness is the best-known and most studied behavioral asymmetry (Papadatou-Pastou, 2011). It can be defined as “the individual’s preference to use one hand predominately for unimanual tasks and/or the ability to perform these tasks more efficiently with one hand” (Corey et al., 2001, p. 144). Handedness seems to be an early developmental characteristic both phylogenetically (i.e., in the evolution of the human species; Corballis, 1991; Lozano et al., 2009) and ontogenetically (i.e., in the development of the individual; Hepper, 2013; Hepper et al., 1998; Rodriguez and Waldenström, 2008). Left-handedness levels in the general population are a point of dispute amongst studies, but large-scale studies provide estimates of around 10% (Perelle and Ehrman, 1994; Peters et al., 2006). This tendency toward right-handedness has been present throughout human history and across cultures and continents (Coren and Porac, 1977; Faurie and Raymond, 2004; Hardyck and Petrinovich, 1977; McManus, 1991, 2009). Importantly, handedness is an overt reflection of the lateralization of the central nervous system. The most extensively demonstrated functional laterality that has been linked with handedness is cerebral lateralization for language (e.g., Bishop et al., 2009; Gonzalez and Goodale, 2009; Isaacs et al., 2006; Khedr et al., 2002; Knecht et al., 2000; Sommer et al., 2002). For example, Knecht et al. (2000) showed, using functional transcanial Doppler sonography (fTCD) in 326 healthy individuals, that the prevalence of right-hemispheric dominance increases linearly with the degree of left-handedness as measured by the Edinburgh Handedness Inventory (Oldfield, 1971), from 4% in strong right-handers, to 15% in ambidextrous individuals, to 27% in strong left-handers. But handedness has also been associated with the cerebral lateralization of a number of other cognitive functions, such as visuospatial attention (e.g., Cai et al., 2013) and embodied cognition (e.g., Willems et al., 2010a,b). Willems et al. (2010a,b) have even demonstrated that the relationship between handedness and hemispheric lateralization extends beyond cognitive functions, to functionally lateralized parts of the visual cortex, which indicates a general coupling between cerebral lateralization and handedness. In a recent review, Willems et al. (2014) presented a bulk of research evidence showcasing that left-handed people show increased rates of reductions or reversals of lateralized brain functions, compared to right-handers, although relationships are in many cases subtle. Brain structural asymmetries have also been linked to handedness. Differences between left- and right-handers have been found for a number of brain structures such as the corpus callosum, the hippocampus, and the amygdala (Anstey et al., 2004; Good et al., 2001; Luders et al., 2003; Sommer et al., 2008; Welcome et al., 2009), as well as the cerebral cortex (Geschwind et al., 2002; Sun and Walsh, 2006). Overall, handedness appears to be an important tool for understanding the normal variation of functional and structural cerebral lateralization. This fact, together with the difficulty in directly assessing cerebral laterality in large and/or special populations (e.g., ID individuals or individuals with autism or learning difficulties) has led to the extensive study of handedness. How handedness interacts with intelligence (or the different cognitive abilities that together manifest themselves as an individual’s general intelligence) is still a matter of ongoing investigation (for a review see Somers et al., 2015). Studies using general population samples have reported reduced levels of cognitive ability in left-handers (e.g., Nicholls et al., 2010, 2012; Resch et al., 1997), elevated levels of cognitive ability in left-handers (e.g., McManus and Mascie-Taylor, 1983), or have failed to find any differences between left- and right-handers (e.g., Hardyck and Petrinovich,

1977; Hardyck et al., 1976). Studies with a focus on degree rather than direction of handedness reported a link between weak relative hand skill asymmetry and elevated levels of cognitive ability (e.g., Annett, 1992; Annett and Manning, 1989) or no advantage in verbal or spatial abilities in participants who had weak hand skill asymmetry, either in children or adults (e.g., Cerone and McKeever, 1999; Mayringer and Wimmer, 2002; McManus et al., 1993; Natsopoulos et al., 2002; Palmer and Corballis, 1996). Other studies (Crow et al., 1998; Leask and Crow, 2001, 2006; Nettle, 2003) including recent large-scale studies (Corballis et al., 2008; Johnston et al., 2009; Peters et al., 2006) corroborate to a link between reduced levels of cognitive ability to mixed-handedness or weak relative hand skill. In addition to the studies using general population samples, several studies have investigated the relationship between handedness and intelligence in intellectually gifted (IG) populations and ID populations. For example, investigations into the members of the Mensa Society, whose membership requirements include possessing an IQ in the top 2% of the world’s population, showed that approximately 20% of the members of Mensa were left-handed (Ehrman and Perelle, 1983; Granville et al., 1979, 1980). Halpern et al. (1998) examined the medical college admission scores for approximately 150,000 adults. They found that left-handers were over-represented among the upper tail on cognitive ability tests. A few more studies that have compared IG individuals with a control group of individuals of normal intelligence have shown that the proportion of left-handers is greater in gifted than in non-gifted individuals (Benbow, 1986; Hicks and Dusek, 1980; Ostatnikova et al., 2002), but others reported the opposite pattern, albeit lacking statistical significance (e.g., Annett, 1993b; Douglas et al., 1967; Piro, 1998). Individuals with ID have also been found to exhibit elevated levels of left-handedness in comparison to the general population both in children (e.g., Carlier et al., 2006; Grouios et al., 1999; Leconte and Fagard, 2006; Pipe, 1987; Pipe and Beale, 1983) and in adults (e.g., Di Nuovo and Buono, 1997; Lewin et al., 1993; Lucas et al., 1989). The prevalence of left- and mixed-handedness was found by some studies to be about twice those reported for control groups of normal intelligence (Lucas et al., 1989; Silva and Satz, 1979; Pipe, 1987). Moreover, it has been found that the proportion of left-handers increases as IQ decreases (Bradshaw-McAnulty et al., 1984; Carlier et al., 2011; Geschwind and Behan, 1982; Gregory and Paul, 1980; Hicks and Barton, 1975; McBurney and Dunn, 1976; Lucas et al., 1989; Pirozzolo and Rayner, 1979; Springer and Eisenson, 1977). By contrast, some studies have failed to report any significant correlation between handedness and IQ (e.g., Annett, 1993b; Barry and James, 1978), while the exact magnitude of the difference in left-handedness between ID individuals and the general population remains under debate. The fact that left-handers are found to be over-represented in the IG and ID populations and yet might not differ from righthanders in their average scores presents no contradiction. The distribution of cognitive ability scores may simply be wider for lefthanders. Nicholls et al. (2010) put forward a different explanation whereby left- and mixed-handedness reflect a shift in the typical pattern of left hemispheric cerebral laterality for hand control and/or language to the right. This re-organization could have either beneficial effects in the interaction with other cognitive functions, leading to enhanced brain function (Benbow, 1986) or deleterious effects, whereby the movement of cognitive functions may cause them to compete for the same neural space, leading to “cognitive crowding” and reduced cognitive ability (Lidzba et al., 2006). The crowding hypothesis was first put forward to explain the paradoxical neuropsychological findings of impaired visuospatial skills and low performance IQ in patients with early left-hemispheric brain damage (Vargha-Khadem et al., 1994; Muter et al., 1997; Lidzba et al., 2006). Since communicative skills are often spared in

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early left-hemisphere lesions by shifting language functions to the right-hemisphere, arising deficits are thought to arise from neuronal scarcity for the functions normally subserved by the right hemisphere. Similar crowding might take place in left- and mixedhanders as cerebral laterality for hand control and/or language is shifted to the right and competes for neuronal space. The shift in the re-organization of cerebral and behavioral laterality has been accounted for by genetic models whereby a gene determines whether or not right-handedness and left-cerebral dominance for language will be expressed (Annett, 1995, 2002; Corballis, 1997; McManus, 1999, 2002). These genetic models typically assume that the genetic variation is preserved through heterozygote advantage (Annett, 1993a). Annett’s (1985) theory, for example, postulates a gene for left cerebral dominance and consequently right-handedness (the right-shift or rs gene) with two alleles (rs− and rs+) and suggests that the heterozygote advantage is such that those individuals with the rs+− genotype have optimal brain organization and hence enhanced cognitive abilities. A situation of balanced polymorphism is thus present, as reproductive success of rs+− is higher than both rs−− and rs++, thereby maintaining the rs− allele in the gene pool, despite the possible cognitive disadvantages associated with the rs−− genotype. McManus (2002), on the other hand, argues that the recessive gene, which causes left-handedness, persists because it is cognitively advantageous. Even though recent findings coming from large-scale twin studies and genome-wide association studies have indicated additive genetic effects for left-handedness (Armour et al., 2014; McManus et al., 2013; Medland et al., 2009; Vuoksimaa et al., 2009), this does not change the rationale of the early genetic models, as the proposed genes of early theories (Annett, 1985; Batheja and McManus, 1985) are hypothetical constructs that might correspond to a group of genes. To date, studies have yielded tentative associations of left-handedness with the genes AR, APOE, COMT, PCSK6, and LRRTM1 (Medland et al., 2005; Francks et al., 2007; Savitz et al., 2007; Bloss et al., 2010; Scerri et al., 2011; Brandler et al., 2013). The elevated levels of left-handedness found in ID samples, have been further explained as cases of “pathological” left-handedness. It has been suggested that left-handedness is associated with a deficit in general cognitive ability because of a brain damage – particularly in the left hemisphere – prenatally or perinatally (Satz, 1972; Satz et al., 1985). Thus, this model predicts that lower cognitive abilities and left-handedness result independently from brain damage. This brain damage is unlikely to account for all lefthanders, but could explain a percentage of cases, differentiating between “normal” and “pathological” left-handers. Indeed, an elevated prevalence of left-handedness has been reported in people who suffered severe bacterial meningitis early in life (Ramadhani et al., 2006) or females with early brain insult (Miller et al., 2005). Moreover, the chances of having a left-handed child are increased for older mothers (Bailey and McKeever, 2004) and for infants who experienced birth asphyxia (Fox, 1985). Yet, a retrospective study of birth complications by Levander et al. (1989) and a large-scale study by Nicholls et al. (2012) have failed to find any association between birth stress and left-handedness, possibly due to the fact that these two studies used medical records, whereas Coren (1995) relied on self-report. It could also be possible that the link between brain pathology and handedness may have a genetic component, as mothers may pass on a susceptibility to problematic births and hence lefthandedness (Pipe, 1987). Coren (1995) found that left-handed mothers were more likely to report having birth-stressed offspring and that left-handers were more likely to have siblings who experienced problematic births. The genetic component of the link between brain pathology and handedness may be further explained in relation to genetically determined neurodevelopmental problems, which might, for example, explain the link between elevated

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levels of atypical handedness and disorders with a neurological component, such as schizophrenia (Sommer et al., 2001; Dragovic and Hammond, 2005) and autism (Hauck and Dewey, 2001). Another hypothesis that has been put forward is that of a maturational delay in ID individuals, which might decrease the degree of right-hand preference or increase the tendency to respond at random (Bishop, 1983, 1990), even though Leconte and Fagard (2006) did not find strong evidence of a delay in the development of handedness in ID children. Leconte and Fagard (2006) suggest that it is rather lack of practice that does not allow right-handedness in ID children to shift from a slight advantage at birth to a stable pattern of preference after a few years. The authors further add that ID children might not be strategic enough to stabilize the right hand as the preferred hand, which would be energy- and time-saving. A number of factors have been put forward to explain the discrepancy between studies’ findings with regards to the exact prevalence of handedness in intellectually gifted (IG) populations and populations with intellectual disability (ID). For example, characteristics of the populations, such as their age or percentages of the two sexes in the sample, may account for the variability in the prevalence of handedness as measured by different studies. Handedness is usually considered to be established around three to seven years of age, even though evidence is not conclusive (Hardyck et al., 1975; McManus et al., 1998; Raymond and Pontier, 2004). A sex difference in handedness, whereby males are 23% more likely then females to be non-right-handed, has been also demonstrated by a large meta-analysis (Papadatou-Pastou et al., 2008; Martin et al., 2011). Age and sex might have a combined effect, as there is evidence that males develop hand preference, or at least righthand preference, later than females (Annett, 1974; Archer et al., 1988). Thus, if the ID or IG samples include a larger (or smaller) percentage of males than the typically developing (TD) samples, then they are likely to produce smaller (or larger) left-handedness scores too. Another factor might be the presence of familial sinistrality (i.e., a familial history of left-handedness), which has been found to interact with individual laterality factors to alter cognitive skills. For example, Mellet et al. (2014) found that performance in verbal and spatial domains was reduced in adults with familial sinistrality, who also exhibited non-maximal preference strength in the dominant hand. Factors pertaining to the design of the studies may be another source of divergence in the literature. Assessment can vary in terms of the instrument used to study handedness, the cut-off criteria used to separate handedness groups, whether direction or degree of handedness is measured, whether hand preference or hand skill is investigated and whether handedness classifications are binary (i.e., left- and right-handers) or include more than two groups (e.g., include an ambidextrous or mixed-handedness group) (for a discussion of these factors see Papadatou-Pastou et al., 2008, 2013). Possible moderation might also reflect secular change, as it has been shown that there is a gradual easing of cultural pressures against sinistrality (McManus and Bryden, 1992). Another factor that could exert moderating effects on the findings of the studies are the methods used to determine intellectual ability, as well as the cut-off points used to separate groups of different intellectual abilities. Overall, the literature on the handedness of populations with ID or IG points to the direction of an increased prevalence of left-handedness amongst both ID and IG populations. Still, not all research findings are in line, nor has the magnitude of this postulated effect been calculated. As discussed, it is possible that the variance between different studies can be explained by their different characteristics, such as the age of the population tested or the handedness classification procedures used. A systematic review and meta-analysis could address these pending issues (Hunter and Schmidt, 1990). A meta-analysis is an analytical technique which integrates and quantifies results from all studies on a clearly defined

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field, thereby overcoming the subjectivity of a conventional literature review (Papadatou-Pastou et al., 2008; Walker et al., 2008). Moreover, in a meta-analysis even studies of small sample sizes can contribute to the final result, heterogeneity among studies can be estimated, ascertainment bias can be investigated, and moderators can be detected. The present study comprises five meta-analyses of studies that have assessed the prevalence of handedness among individuals of different IQ levels – ID, TD, and IG individuals. The main goal of the present meta-analysis is to provide a definitive test of the hypothesis that there is a higher prevalence of atypical handedness among IG and ID individuals compared to TD individuals and to estimate the overall magnitude of this difference. Further, another goal is to estimate whether heterogeneity among studies exists and, if so, which are the sources of it. Potentially moderating factors include the IQ test employed, the handedness classification used, and the year of publication of the study. In addition to the above, the metaanalyses will directly assess whether ascertainment bias has played a role in the differential reporting of the left-handedness prevalence amongst individuals of different IQ levels. 2. Method The studies that were entered into the meta-analyses were located via online databases Pubmed MEDLINE at PUBMED (NLM) (274 articles) and PsychINFO (97 more articles) using the terms (handedness OR hand preference OR hand skill) AND (intelligence OR IQ). Reference lists were also scanned (26 more articles). The number of the abstracts screened was 397 (299 articles were excluded at this stage). The number of full text articles assessed for eligibility was 98. Sixteen studies were included in the metaanalyses. Data collection ended in May 2015. 2.1. Study selection The following criteria were set for inclusion of an individual study in the meta-analyses. 1. Control group: Studies were required to include a control group of TD individuals, otherwise they were not included in the analyses (e.g., Di Nuovo and Buono, 1997; Storfer, 1995). Two studies were excluded (Benbow, 1986; Wiley and Goldstein, 1991) because their control group was characterized as “less gifted”, thus it was not clear if it comprised of TD individuals. It should be noted that we included the study of Porac et al. (1980), which reports two controls groups, but only the control group that was matched to the ID group on chronological age and not mental age was included in the analysis, to ensure consistency with the other studies. 2. Absence of comorbidity: In studies on individuals with ID, the disability had to be of unknown or idiopathic nature. Accordingly, studies including participants diagnosed with Down’s syndrome, attention deficit hyperactivity disorder (ADHD), or who had low birth weight were excluded (e.g., BradshawMcAnulty et al., 1984; Ross et al., 1992; Segal, 1989). Still, the study of Pipe and Beale (1983) was included, although three out of the twelve ID children had Down’s syndrome, as none had gross motor or sensory defects (analyses were also repeated excluding this study). 3. Publication language: Reports written in English were included. The only exception was the unpublished thesis of the second author (Tomprou, 2013). 4. Studies whose participants were selected on the basis of their handedness in order to increase the percentage of left-handed participants were carefully excluded (e.g., Annett and Turner,

1974; Natsopoulos et al., 1992; Ghayas and Adil, 2007; Resch et al., 1997). 2.2. Moderators The variables whose possible moderating effects were examined included the following: Year of publication. For investigating possible moderation in terms of secular change, the year of publication was entered numerically for each study. Year of publication of each study has been previously used as a proxy for secular change (e.g., PapadatouPastou et al., 2008). Classification of handedness. The classification of handedness in the studies that were included in the meta-analysis followed either a binary classification with two handedness classes (RH-LH, RH-nonRH, or LH-nonLH) or a classification with three handedness classes (RH-LH-MH, RH-LH-inconsistent, RH-LH-ambilateral, or RH-LH-no preference binary). The studies were coded for classification of handedness using two different groupings: (a) studies with two handedness classes and (b) studies with three handedness classes. In the case of the Barry and James (1978) study where two different cut-off points were used for the determination of the handedness groups, the cut-off points suggested by Annett (1967) were used instead of those suggested by Colby and Parkison, 1997, because Annett uses “more stringent cutoffs for right- and lefthandedness” (Barry and James, 1978, p. 320) and because the classification of Colby and Parkison (1997) “has a number of problems associated with it, not least of which is the arbitrary assignment of a 60% cutoff figure for laterality categorization” (Barry and James, 1978, p. 316). Intelligence Measurement. Intelligence was mainly measured by tests of mental ability, but in some studies IQ was based on official or school records. For example in Batheja and McManus (1985) the ID children attended Educationally Subnormal (Severe) schools, while the TD children attended normal schools. The IQ tests comprised a battery of mental tests, the Progressive Matrices of Raven (Raven, 1965; Raven, 1976), the Wechsler Intelligence Scale for Children (WISC) (Wechsler, 1974; Wechsler, 1991) or the Stanford–Binet Intelligence Scales (Thorndike et al., 1986). Thus, the studies were coded for instrument using two different groupings: (a) official or school records and (b) IQ tests. Sex ratio. Possible sex differences in the putative effect could not be directly investigated, as the included studies did not break down their results by sex. The sex ratio of the participants was therefore used and was entered numerically as the percentage of male participants. Six studies did not report the sex distribution of their samples at all (Douglas et al., 1967; Gordon, 1921; Hicks and Dusek, 1980; Piro, 1998; Searleman et al., 1988; Wilson and Dolan, 1931) and could not be entered into the sex ratio analysis. Some further variables with a possible moderating effect were not examined due to missing or unusable data, namely age, degree of handedness, instrument used to measure handedness, as well as whether hand preference or hand skill was assessed. In the case of age, one study did not report the range or mean age of participants at all (sample age was indicated as “junior high school pupils” for the TD group and “6-A pupils” for the ID group, Wilson and Dolan, 1931, p. 261). The remaining studies reported mean IQ scores of their participants groups, without breaking them down in different age groups, with the exception of Piro (1998), where participants’ data were broken down in two age groups (8–10 and 11–14 years old; in the present analysis they were treated as a single data set). In addition, most studies used wide age ranges that were overlapping for the most part (e.g., 7–18 years. old, Batheja and McManus, 1985, 7–14 years. old, Ostatnikova et al., 2002), thereby not allowing for the effect sizes of different age groups to be compared. When it came to the instrument used to measure handedness, with

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the exception of three studies using the Edinburgh Handedness Inventory by Oldfield (1971; Ostatnikova et al., 2002; Pickersgill and Pank, 1970; Tomprou, 2013), all the other studies used handedness measurements that were different (at most two studies were similar), not allowing for meaningful groups to be formed. Only two studies reported sufficient data from a hand skill test (Annett, 1993a,b; Tomprou, 2013) and only six reported data on the degree of handedness (Barry and James, 1978; Batheja and McManus, 1985; Leconte and Fagard, 2006; Pickersgill and Pank, 1970; Pipe and Beale, 1983; Tomprou, 2013). 2.3. Statistical analysis Data were analyzed using the Comprehensive Meta-Analysis software package (v. 2.; Borenstein et al., 2005). Two sets of metaanalyses were conducted; one comparing handedness prevalence between ID and TD individuals and another set comparing handedness prevalence between IG and TD individuals. Moreover, within each set, two different meta-analyses were carried out, based on the different handedness classifications employed by different researchers: (a) left-handedness, where left-handers were either (i) the participants classified as left-handers in two of the binary classifications (RH-LH or LH-nonLH) but not the RH-nonRH classification or (ii) the participants at the left-extreme of three-way classifications (RH-LH-MH, RH-LH-inconsistent, RH-LH-ambilateral, or RH-LH-no preference binary) and (b) non-right-handedness, where non-right-handers were either (i) the participants classified either as left-handers or non-right-handers in the studies using binary classifications or (ii) the participants that were classified as leftextreme or MH, inconsistent, ambilateral, or having no preference in studies employing three-way classifications. In the first four meta-analyses, the measure of effect size used was the odds ratio (OR), which is defined as the ratio of the odds of an event occurring in one group to the odds of it occurring in another group. In this case, the event would be left-handedness (or non-right-handedness) and the two groups would be the ID and TD individuals for the first set of meta-analyses and IG and TD individuals for the second set of meta-analyses. The main advantage of the OR is that it is independent of the base rate of the event in question (i.e., left-handedness or non-right-handedness) within each study. ID to TD (or IG to TD) ORs, as well as corresponding two-tailed 95% confidence intervals (95% CI) were calculated for each data set independently. An overall effect (combined effect) was also calculated using a fixed effects model to provide a pooled OR and a test for the overall effect (Z statistic). An OR value of 1.0 corresponds to the null hypothesis of no differences in handedness levels among individuals with different IQ levels, whereas values greater than 1.0 indicate a larger proportion of atypical handedness (left- or nonright-handedness) among individuals with ID (or IG for the second set of meta-analyses). Though the OR has mathematical properties which make it particularly suitable for combining evidence, it is generally recognized that simple proportions may be easier to grasp intuitively. It can be shown that the relationships investigated in the present study take the form of ID = TD × OR/[1 + TD(OR − 1)], where ID and TD are the probabilities of atypical handedness in ID and TD populations, respectively. (In the place of ID, IG, the probability of atypical handedness in IG populations, can also be used.) Moreover, each comparison was tested for heterogeneity using the homogeneity statistic Cochran’s Q. The Q statistic is used to ascertain whether the participants in different studies originate from the same or different populations. The extent of inconsistency among the data sets’ results was assessed using the I2 index. The I2 index can 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). Higgins et al.,

155

2003 have proposed that I2 index levels of 25%, 50%, and 75% may be described as low, moderate, and high, respectively. In the case of non-significant heterogeneity between the data sets the fixed effect model was used, which assumes that all the data sets included in the meta-analysis come from a single population and asks what the best estimate of the true effect size of the population is (in the present case the odds ratio). Otherwise, in the case of statistically significant heterogeneity, the analysis was repeated using a random effects model, which assumes that the included data sets are drawn from a distribution of populations and asks what the range and distribution of odds ratios in the sample of populations studied is. This two-step procedure has been coined “conditionally random-effects” procedure by Hedges and Vevea (1998). For both models, differences in the effect sizes across the data sets can be attributed to sampling error, but only for the random-effects model the between-study heterogeneity is due to moderating factors, such as individuals’ characteristics, methodology of the study or instruments used for data collection. Forest plots were used to depict all the information visually. The data sets were also tested for ascertainment bias using the funnel plot graphical test (Light and Pillemer, 1984), Egger’s t statistical test (Egger et al., 1997), Duval and Tweedie, 2000 trim and fill method and the fail-safe N (Rosenthal, 1979). The rationale behind the funnel plot is that in the absence of ascertainment bias, standard errors of the studies are distributed symmetrically around the overall effect size and therefore one should expect a symmetrical funnel plot. Egger’s t test is a regression-based test that estimates asymmetry of the funnel plot, with positive values of the intercept a (a > 0) indicating a trend toward higher levels of test accuracy in studies with smaller sample sizes. Duval and Tweedie (2000) trim and fill method of correcting bias is a nonparametric (rank-based) data augmentation technique. It is a two-step method: in the trimming phase, small studies are excluded in order for the plot to acquire symmetry and then an adjusted summary effect is estimated considering only the larger studies. In the filling phase, the funnel plot is replicated by replacing the excluded studies with their missing counterparts around the adjusted summary estimate. The fail-safe N is the number of data sets with an odds ratio of one that would be needed to be added to the existing meta-analysis so that the overall effect size is no longer significant at the level of p < .05. The possible moderating effects of the interval moderator variables (i.e., year of publication and sex ratio) was examined using meta-regression. Random effects models were used as recommended by Thompson and Higgins (2002), with evaluation in terms of the Q statistic. In the cases of the categorical moderator variables (i.e., handedness classification and IQ instrument) the effect sizes in the different subgroups that form the levels of each moderator were again compared by means of the Q statistic. A separate, fifth meta-analysis was conducted using the four studies that reported means as well as standard deviations on the degree of handedness (Barry and James, 1978; Leconte and Fagard, 2006; Pipe and Beale, 1983; Tomprou, 2013) using the standardized difference of means as the effect size.

3. Results A total of 16 studies were included in the analyses, comprising 19 separate data sets (13 data sets comparing ID and TD individuals and 6 data sets comparing IG and TD individuals) totaling 16,076 individuals (5795 ID individuals, 8312 TD individuals, and 1969 IG individuals). The details of all the studies used can be found in Table 1. Two sets of analyses were conducted, one on the data sets comparing ID to TD individuals, and then a second one comparing IG to TD individuals. A separate analysis was conducted using the four studies that included information on the degree of handedness.

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Table 1 Studies included in the meta-analysis on intelligence and handedness. Study

N (total)

N (Intellectually Deficient)

N (Typical IQ)

N (Intellectually Gifted)

Age

Handedness Instrument

Classification of Handedness

Intelligence Instrument

Classification of Intelligence

Atypical Handedness (Mixed & Left)/Left Handedness %

Notes

Mentally defectivenormal Regularspecial

13.69

18.24

7.30

–

4.28

10.59

3.54

–

No data regarding number of males and females participants. No exact data regarding mean or range of age, as well as number of males and females participants.

7918

4620

3298

–

8–16

Battery of handedness tests

LH-RH

Official records

a, Wilson and Dolan

1448

151

1297

–

Junior high school pupils

Indication of preferred hand

LH-RH

Battery of IQ tests

3290

457

2168

665

8–15

School doctors’ records

LHinconsistentRH

Battery of IQ tests

1 SD below and mentally handicappedmean score −1 SD above

15.44/5.78

20.79/6.13

14.80/5.90

13.83/5.11

24

8 (7 boys)

16 (11 boys)

–

10.9–17.1

Edinburgh Handedness Inventory (EHI)

LH-RH

Official records

Subnormal/non mongolnormal

25.00

37.50

18.75

–

68

34 (30 boys)

34 (30 boys)

–

4.1–18.11

Colby and Parkison (1997) battery of handedness tasks

LHAmbilateralRH

School records

Retardednormal

54.41/4.41

52.94/2.94

55.88/5.88

–

969

–

391

578

8–12

Briggs-Nebes Scale (1975)

LH-MH-RH

Stanford Binet

Gifted-non gifted

17.85/8.36

–

17.13/9.97

18.34/7.27

309

138 (76 boys)

555 (278 boys)

–

14–24

4-item handedness performance measure

LH-MH-RH

Official records

Retardatesnormal

37.54/10.36

60.14/15.94

19.30/5.85

–

24

12 (6 boys)

12 (6 boys)

–

5.3–18.1

Annett’s Handedness Index (7-item)

nonRH-RH

Stanford Binet/Peabody MA

Retardednormal

37.50

41.67

33.33

–

92

45 (26 boys)

47 (27 boys)

–

7–18

10-item handedness test

LH-RH

Official records

Non Down’snormal

18.48

26.67

10.64

–

302

90

212

–

16–22

4 unimanual tasks

LH-no preference-RH

Official records

Retardednormal

19.87/15.89

28.89/23.33

16.04/12.74

–

342

107 (43 boys)

180 (94 boys)

55 (32 boys)

5.1–11.8

Observation of 7 actions

LH-MH-RH

Raven CPM

Dull-averagebright

33.04/3.22

31.78/1.87

33.89/3.89

32.73/3.64

Data for hand skill (peg moving test) were also provided.

b, Piro (1998)

618

–

212

406

8–14

LH-MH-RH

14.08/10.19

–

16.04/14.15

13.05/8.13

146

73 (43 boys)

73 (43 boys)

–

13.9–18.1

WISC-R, WISC-III, Stanford Binet Raven CPM

Gifted-non gifted

a, Grouios et al.

Dean Laterality Preference Schedule (DLPS) Briggs and Nebes’s 12-item Handedness Inventory

Intellectual disabilitynormal

34.93/13.70

56.16/17.80

13.70/9.59

–

No data regarding number of males and females participants. Intellectual ability of normal children was based on teacher reports.

(1931)

a ,b, Douglas et al. (1967)

a, Pickersgill and Pank (1970)

a ,c, Barry and James (1978)

b, Hicks and Dusek (1980) a, Porac et al. (1980)

a ,c, Pipe and Beale (1983)

a, Batheja and McManus (1985)

a, Searleman et al. (1988)

a ,b Annett (1993a,b)

(1999)

LH-MH-RH

Data from mentally handicapped and children with 1 SD below average intelligence were integrated. No data regarding number of males and females participants. Data were also provided for children with mongolism, as well as adults. Two handedness classification criteria (in the present meta-analysis was chosen Annett’s classification, 1967). Data were also provided for children in the autistic spectrum. No data regarding number of males and females participants. Although both retardates and normal group included adults up to 24 years, the mean age for both groups was 17 years and 16.8 years respectively. Three children had Down Syndrome. Controls and retarded children were matched according to their mental age and not chronological as in most studies. Data from a Down syndrome group were provided, but they were carefully excluded. Although some participants were adults, 92% was under 19 years old. No data regarding number of males and females participants.

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a, Gordon (1921)

–

Data from another handedness test (Quantification of Hand Preference Test) and a hand skill test (Peg moving) were also provided. 21.42

13.33/13.33

33.33

23.33/3.33

c

a

b

30 (18 boys) 42 (21 boys) 30 (17 boys) 102 (2013)

a ,b , c, Tomprou

Studies included in the ID-TD meta-analyses. Studies included in the IG-TD meta-analyses. Studies included in the ID-TD degree of handedness meta-analysis.

13–17

Edinburgh Handedness Inventory (EHI)

LH-MH-RH

WISC-III

Intellectual deficienttypical IQ-intellectual gifted

27.45

20.00/6.67 Intellectually deficienttypically developing Raven CPM LH-MH-RH 10–14 – 15 (8 boys) 30 (19 boys) 45 Fagard (2006)

a ,c, Leconte and

157

3.1. Results of the ID-TD analyses

30.00

Data from a card reaching task were provided. Two groups of intellectually deficient children were examined: one age matched to controls and one 2 years older.

– 6.38 4.17 – 5.54 Intellectual gifted-average WISC-III, Stanford Binet LH-nonLH

Edinburgh Handedness Inventory (EHI) 12-item Laterality Test 7–14 235 (144 boys) 144 (63 boys) – 379 (2002)

b, Ostatnikova et al.

Study

Table 1 (Continued)

N (total)

N (Intellectually Deficient)

N (Typical IQ)

N (Intellectually Gifted)

Age

Handedness Instrument

Classification of Handedness

Intelligence Instrument

Classification of Intelligence

Atypical Handedness (Mixed & Left)/Left Handedness %

Notes

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Left-handedness. This comparison included k = 12 studies adding up to n = 15,167 individuals (n = 6746 ID individuals, n = 8421 TD individuals). Fixed effects analysis gave a pooled odds ratio (OR) = 2.55, 95% confidence interval (CI) = 2.24, 2.91, Z = 14.18, p < .001 (see Fig. 1). Significant heterogeneity was found to exist among the data sets, Q (11) = 47.93, p < .001, with high inconsistency between studies, I2 = 77.05%, indicating that one or more variables may moderate the relationship between low IQ and handedness. A random effects model was therefore employed, which revealed that individuals with ID seemed to have a greater tendency toward left handedness compared to TD individuals, OR = 1.98, 95% CI = 1.24, 3.15, indicating that the OR was significantly different from 1.0 (Z = 2.87, p = .004). Thus, if the prevalence of TD left-handedness were exactly 10%, the observed estimate of 1.98 for the ratio of ID to TD individuals left-handedness odds would be equivalent to a prevalence of ID left-handedness of 19.6%. Due to the large sample size of one study (n = 4620 ID individuals, n = 3298 TD individuals; Gordon, 1921), a further analysis was conducted without this particular study. The comparison included k = 11 studies adding up to n = 7249 individuals (n = 2126 ID individuals, n = 5123 TD individuals). Fixed effects analysis gave a pooled OR = 1.91, 95% CI = 1.48, 2.46, Z = 5.02, p < .001. Significant heterogeneity was found to exist among the data sets, Q (10) = 41.12, p < .001, with high inconsistency between studies, I2 = 75.68%, indicating that one or more variables may moderate the relationship between low IQ and handedness. A random effects model was therefore employed, which revealed that individuals with ID seemed to have a greater tendency toward atypical handedness compared to TD individuals, OR = 1.73, 95% CI = .94, 3.18, indicating that the OR was not significantly different from 1.0 (Z = 1.75, p = .08). Therefore, excluding the Gordon (1921) study did not make the sample homogeneous, hence a random effects model was employed. The ratio of ID to TD individuals left-handedness odds is now numerically smaller, going from 1.98 to 1.73. Importantly, the exclusion of the Gordon (1921) study changed the statistical significance of the results. The results again show that ID individuals tend to present higher levels of atypical handedness compared to TD individuals, but this difference is not statistically significant. A second reanalysis was performed without the Porac et al. (1980) study, because the OR of the study (OR = 10.3) is an outlier (the second largest outlier is 3 times smaller at OR = 3.22 for the Douglas et al., 1967 study). The comparison included k = 11 studies adding up to n = 14,858 individuals (n = 6608 ID individuals, n = 7.866 TD individuals). Fixed effects analysis gave a pooled OR = 2.45, 95% CI = 2.15, 2.80, Z = 13.38, p < .001. Significant heterogeneity was found to exist among the data sets, Q (10) = 35.04, p < .001, with high inconsistency between studies, I2 = 71.46%, indicating that one or more variables may moderate the relationship between low IQ and handedness. A random effects model was therefore employed, which revealed that individuals with ID seemed to have a greater tendency toward atypical handedness compared to TD individuals, OR = 1.67, 95% CI = 1.07, 2.62, indicating that the OR was significantly different from 1.0 (Z = 2.25, p = .02). Therefore, when excluding the Porac et al. (1980), the ratio of ID to TD individuals left-handedness odds went down to 1.67 from 1.98, but remained significant. Non-right-handedness. This comparison included k = 13 data sets adding up to n = 15,725 individuals (n = 6988 ID individuals, n = 8737 TD individuals). Fixed effects analysis gave a pooled OR = 2.62, 95% CI = 2.34, 2.94, Z = 16.72, p < .001 (see Fig. 2). Significant heterogeneity was found to exist among the data sets, Q (12) = 128.01, p < .001, with very high inconsistency between studies, I2 = 90.63%. A random effects model was therefore employed,

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Fig. 1. Forest plot of the intellectually disabled (ID) to typically developing (TD) children odds ratios for the left-handedness comparison. In the plot the 95% confidence interval for each study is represented by a horizontal line and the point estimate is represented by a square. The confidence intervals for totals are represented by a diamond shape at the bottom of the plot.

Fig. 2. Forest plot of the intellectually disabled (ID) to typically developing (TD) children odds ratios for the non-right-handedness comparison.

which revealed a clear difference in non-right-handedness between individuals with ID and TD individuals, OR = 2.66, 95% CI = 1.63, 4.35, indicating that the OR was significantly different from 1.0, Z = 3.90, p < .001. Thus, if the prevalence of TD non-righthandedness were exactly 10%, the observed estimate of 2.66 for the ratio of ID to TD individuals non-right-handedness odds would be equivalent to a prevalence of ID non-right-handedness of 31.02%. Again, a further analysis was conducted without the study of Gordon (1921) and the fixed effects analysis gave a pooled OR = 2.38, 95% CI = 2.01, 2.82, Z = 9.95, p < .001. Significant heterogeneity was found to exist among the data sets, Q (11) = 125.75, p < .001, with very high inconsistency between studies, I2 = 91.25%. A random effects model was therefore employed, which revealed a clear difference in non-right-handedness between individuals with ID and TD individuals, OR = 2.60, 95% CI = 1.33, 5.09, indicating that the OR was significantly different from 1.0, Z = 2.78, p = .005.2 The

results are almost identical to the ones computed including the Gordon (1921) study, with the ratio of ID to TD individuals nonright-handedness odds now only going down to 2.60 from 2.66. A second reanalysis was again conducted without the study of Porac et al. (1980), which was again an outlier with an OR = 23.87. The fixed effects analysis gave a pooled OR = 2.32, 95% CI = 2.06, 2.60, Z = 14.17, p < .001. Significant heterogeneity was found to exist among the data sets, Q (11) = 45.50, p < .001, with high inconsistency between studies, I2 = 75.83%. A random effects model was therefore employed, which revealed a clear difference in non-right-handedness between individuals with ID and TD individuals, OR = 2.11, 95% CI = 1.49, 2.98, indicating that the OR was significantly different from 1.0, Z = 4.20, p < .001. Therefore, when excluding the Porac et al. (1980), the ratio of ID to TD individuals non-right-handedness odds went down to 2.11 from 2.66, but remained significant.

2 All analyses for this comparison were also repeated without the Pipe and Beale (1983) study, in order to make sure that the inclusion of the three participants with Down’s Syndrome did not alter the results. In this case, the fixed effects analysis gave a pooled OR = 2.66, 95% CI = 2.35, 2.95, Z = 16.73, p < .001, Q (11) = 127.50, p < .001, I2 = 91.37%. The random effects model gave an OR = 2.74, 95%, CI = 1.65, 4.55, Z = 3.90,

p < .001. Excluding Gordon (1921), the fixed effects analysis gave a pooled OR = 2.39, 95% CI = 2.01, 2.84, Z = 9.96, p < .001, Q (10) = 125.39, p < .001, I2 = 92.03%. The random effects model gave an OR = 2.71, 95% CI = 1.34, 5.46, Z = 2.78, p = .005.

M. Papadatou-Pastou, D.-M. Tomprou / Neuroscience and Biobehavioral Reviews 56 (2015) 151–165

159

Fig. 3. Funnel plot of standard error on log odds intellectually disabled (ID) to typically developing (TD) children ratio, for the non-right-handedness comparison.

Table 2 ID children to TD non-right-handedness odds ratios (OR) and 95% confidence intervals (CI) for all levels of the moderator variables. Moderator variable

Level

Number of data sets

OR

Classification of Handedness

Two classes (LH-RH, non RH-RH, LH-nonLH) Three classes (LH-MH-RH, LH-inconsistent-RH, LH-ambilateral-RH, LH-no preference-RH) Official/school records IQ tests

5 8

2.84 2.31

2.46–3.28 1.07–6.35

6 7

3.33 2.10

1.31–8.44 1.22–3.63

Intelligence Instrument

The data sets that were included in the non-right-handedness comparison were then tested for ascertainment bias. This comparison included n = 15,725 individuals (n = 6988 ID analysis, n = 8737 TD analysis) from all 13 data sets and is therefore the most representative one. No ascertainment bias was detected using Egger’s Test, t (11) = .07, p = .95. Similarly, visual inspection of the funnel plot graphical test did not provide evidence of asymmetry and hence possible ascertainment bias (see Fig. 3). Using Duval and Tweedie’s trim and fill method for bias correction for the random effects model no data sets were “trimmed” or “filled”, meaning that an adjusted odds ratio needed not be calculated. Finally, the fail-safe N was calculated, and found to be N = 524. The value of N indicates that 524 unpublished studies with an odd ratio 1.0 (which means no handedness differences among ID and TD individuals), would be needed to be added to the existing meta-analysis so as for the results to be non-significant. The high value of N even exceeds the

95% CI

number of the studies included in meta-analysis and confirms the reliability of the observed effect. Because of the heterogeneity detected among studies, the moderating effects of the previously indicated variables were tested within the non-right-handedness comparison (see Table 2):

1. Publication year. Meta-regression of the year of publication of the studies was conducted and revealed no significant linear trend on the size of the ID-to-TD odds ratio in non-right-handedness, Q (1) = 1.37, p = .24 (see Fig. 4). 2. Sex ratio. Meta-regression of the sex ratio of the studies was conducted and revealed a trend on the size of the ID-to-TD odds ratio in non-right-handedness, Q (1) = 3.46, p = .06 (see Fig. 5), but which disappeared if the outlying study of Porac et al. (1980) was not included in this analysis, Q (1) = 0.07, p = 0.78.

Fig. 4. Meta-regression of the year of publication of the data set on log odds intellectually disabled (ID) to typically developing (TD) children ratio, for the non-right-handedness comparison.

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Fig. 5. Meta-regression of the sex ratio of the data set on log odds intellectually disabled (ID) to typically developing (TD) children ratio, for the non-right-handedness comparison.

3. Classification of handedness. The odds of being non-right-handed were found to be 184% and 131% higher for ID than TD individuals in two-way and three-way classifications, respectively (see Table 2). Nevertheless, the moderating effect of classification was not found to be significant, Q (1) = .04, p = .85. The confidence intervals overlap, with the 95% CI of the three-way classifications (1.07, 6.35) containing the 95% CI (2.46, 3.28) of two-way classifications. Heterogeneity was further examined within the two classifications, which revealed significant heterogeneity only within the three-way classification, Q (7) = 124.09, p < .001, I2 = 94.36%, but not within the two-way classification, Q (4) = .86, p = .93, I2 = 0%. The three-way classification could not be broken down any further, though, as the number of studies was small and the cut-off points used in the different studies to determine the three groupings were varied and could not be used to construct meaningful subgroups. 4. Intelligence instrument. The moderating effect of the instrument used to measure intelligence did not reach significance either, Q (1) = .70, p = .40. The odds of being non-right-handed were found to be 233% higher for ID than TD individuals when official/school records were used to measure intelligence and 110% when standardized IQ tests were used (see Table 2). Again, confidence intervals are overlapping, with 95% CI = 1.31, 8.44 for official/school records and 95% CI = 1.22, 3.63 for IQ tests. Heterogeneity was further examined within the two measurement types, which revealed significant heterogeneity both within the studies using official/school records to measure intelligence, Q (5) = 76.03, p < .001, I2 = 93.42%, and within the studies using standardized IQ tests, Q (6) = 25.28, p < .001, I2 = 76.26%. Again, meaningful groups for creating sub-groups for further analysis could not be created, due to the variety of measures and tests used and the small number of studies. Overall, no moderating effects were detected for year of publication, classification of handedness, sex ratio, or the instrument used to measure intelligence. As stated above, only six studies reported data on the degree of handedness (Barry and James, 1978; Batheja and McManus, 1985; Leconte and Fagard, 2006; Pickersgill and Pank, 1970; Pipe and Beale, 1983; Tomprou, 2013). The reported data could not be incorporated in the moderating variables analysis, as they were means of handedness scores for each group (ID or TD). Therefore, meaningful groups could not be constructed and compared. Yet, four of these studies (Barry and James, 1978; Leconte and Fagard, 2006; Pipe and Beale, 1983; Tomprou, 2013) further reported the standard deviations of the means for each group, allowing for

a separate meta-analysis to be conducted, this time using the standardized difference of means as the effect size and not the odds ratio. This comparison included k = 4 studies adding up to n = 195 individuals (n = 92 ID individuals, n = 103 TD individuals). Fixed effects analysis gave a pooled standardized differences in means d = −.09, 95% CI = −.37, .19, Z = −.60, p = .55. The heterogeneity among the data sets was not found to be significant, Q (3) = 5.67, p = .13, thus a further analysis based on the random effects model was not necessary. Moreover, since under the fixed effect model all data sets describe a single population, no moderator variables analysis was granted here. 3.2. Results of IG-TD analyses Left-handedness. This comparison included k = 6 data sets adding up to n = 5451 individuals (n = 2098 IG individuals n = 3353 TD individuals). Fixed effects analysis gave a pooled OR = .76, 95% CI = .60, .97, Z = −2.22, p = .03, which revealed that IG individuals seemed to have a greater tendency toward typical handedness compared to TD individuals (see Fig. 6). Thus, if the prevalence of TD left-handedness were exactly 10%, the observed estimate of 0.76 for the ratio of IG to TD individuals left-handedness odds would be equivalent to a prevalence of IG left-handedness of 7.79%. Significant heterogeneity was not found to exist among the data sets, Q (5) = 4.39, p = .49, thus an analysis based on the random effects model was not necessary and no moderator variables analysis was granted either. Non-right-handedness. This comparison included k = 6 data sets adding up to n = 5897 individuals (n = 2262 IG individuals, n = 3635 TD individuals). Fixed effects analysis gave a pooled OR = .97, 95% CI = .82, 1.16, Z = −.31, p = .76 which revealed that IG individuals seemed to have 3% less chance to have atypical handedness, but this difference did not reach statistical significance (see Fig. 7). Thus, if the prevalence of TD non-right-handedness were exactly 10%, the lack of significance of the OR would be equivalent to a prevalence of IG non-right-handedness again of 10%. No heterogeneity was found among the data sets, Q (5) = 3.08, p = .69, thus neither an analysis based on the random effects model nor a moderator variables analysis were performed. Moreover, the data sets of the non-right-handedness comparison were tested for ascertainment bias. No ascertainment bias was detected using Egger’s Test, t (4) = 1.31, p = .26. Visual inspection of the funnel plot graphical test did not provide evidence of asymmetry and hence possible ascertainment bias (see Fig. 8). Using Duval and Tweedie’s trim and fill method for bias correction, for

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Fig. 6. Forest plot of the intellectually gifted (IG) to typically developing (TD) children odds ratios for the left-handedness comparison.

the random effects model, one data set was “filled” and the adjusted odds ratio was OR = .96, 95% CI = .81, 1.14, an estimation almost identical to the OR originally calculated (OR = .97, 95% CI = .82, 1.16). The fail-safe N was not calculated, as the existing meta-analysis was already non-significant.

4. Discussion Five separate meta-analyses of studies on 16,076 individuals (5795 intellectually disabled [ID] individuals, 8312 typically developing [TD] individuals, and 1969 intellectually gifted [IG] individuals) were carried out. The first set of meta-analyses included data sets from studies investigating the handedness prevalence in ID individuals compared to TD individuals, while the second set was concerned with the handedness prevalence of IG individuals compared to TD individuals. Tests of ascertainment bias revealed that there was no evidence that the studies in the two sets had been distorted by preferential reporting. Combined odds ratios for the first comparison provided an estimate of 1.98 for the ratio of ID to TD individuals left-handedness odds (95% CI = 1.24, 3.15), corresponding to a prevalence of 19.6% and an estimate of 2.66 for non-right-handedness odds (95% CI = 1.63, 4.35), corresponding to a prevalence of 31.02%. The estimate for the IG/TD comparison was OR = .76 (95% CI = .60, .97) for the ratio of IG to TD individuals left-handedness odds, corresponding to a prevalence of 7.79% and a non-significant OR = .97 (95% CI = .82, 1.16) for the non-righthandedness odds, meaning that the same prevalence is found for both IG and TD individuals. We argue that these findings can be attributed to cognitive crowding due to a re-organization in the typical pattern of left cerebral laterality for language and hand control to the right and/or to pathological factors.

Over the whole sample, a large proportion of the variability across studies comparing the handedness levels of ID and TD individuals was found to be due to genuine heterogeneity rather than to change. (Of note, the left-handedness comparison was no longer found to present significant levels of heterogeneity after the exclusion of the largest study which included data on 7918 participants; Gordon, 1921). Despite the large levels of variation, none of the three variables whose moderating effect was investigated (year of publication, sex ratio, classification of handedness, and intelligence instrument) were found to exert significant moderating effects. This finding may seem counter-intuitive. Nonetheless, in examining meta-analytic data for effects of moderator variables the crucial characteristic is the number of data sets (13 in this case for the non-right-handedness comparison) and not the number of participants (15,725). This can result in a low power of meta-analytic studies when it comes to examining moderating effects, despite the large numbers of participants (Hunter and Schmidt, 1990). Therefore, with the sample size of the present data set, only very large effects could have been detected. Further variables with a possible moderating effect, such as age, sex, the instrument used to measure handedness, whether hand preference or hand skill was assessed as well as the degree of handedness could not examined within the present meta-analysis because of missing or unusable data. No heterogeneity was found to exist among the data of the IG/TD comparison, thus a moderator variables analysis was not performed for this comparison. The present findings make a strong case for a higher prevalence of atypical handedness among ID individuals compared to TD individuals. Still, the prevalence of atypical handedness is estimated to be between 19.6% and 31.02%. Thus, a word of caution is necessary, making clear that atypical handedness should not be considered a symptom of intellectual disability, as the clear majority of ID indi-

Fig. 7. Forest plot of the intellectually gifted (IG) to typically developing (TD) children odds ratios for the non-right-handedness comparison.

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Fig. 8. Funnel plot of standard error on log odds intellectually gifted (IG) to typically developing (TD) children ratio, for the non-right-handedness comparison.

viduals are right-handers. In the case of IG individuals, only the OR calculated for the left-handedness comparison reached statistical significance (with the upper end of the 95% CI very close to 1 [.97] that represents no difference between IG and TD individuals in lefthandedness levels), but the non-right-handedness comparison was non-significant. These findings constrain the range of acceptable theories on the handedness distribution for different intelligence levels. For example, the suggestion that the distribution of cognitive ability scores may be wider for left-handers is not supported, as elevated levels of left-handedness were not found at both ends of the intelligence spectrum, but were limited to the ID individuals. On the other hand, the suggestion that cerebral reorganization that results from a shift in the typical pattern of cerebral laterality for language and hand control to the left (Nicholls et al., 2010) may result in cognitive crowding and reduced cognitive ability (Lidzba et al., 2006), but not in enhanced brain function (Benbow, 1986; but also see O’Boyle and Benbow, 1990), is in line with the present findings. Cerebral organization of intellectually disabled individuals in the absence of comorbidity such as Down’s syndrome, epilepsy or autism has received surprisingly little research attention in the brain imaging literature. The vast majority of studies that have investigated this question have used handedness as a proxy (i.e., the studies that are included in the present meta-analyses). A study that examined blood flow response to auditory stimulation using transcranial Doppler ultrasonography found a less asymmetrical pattern of blood flow in ID compared to TD children (Bruneau et al., 1992), supporting the idea of a shift of language functions to the right hemisphere in ID individuals. The present findings have limited implications for the genetic explanations of handedness. Annett’s (1985) theory predicts a heterozygote advantage for the r+− genotype, which would be manifested by a weak hand preference. This prediction could not be properly evaluated in the present study, as only a small scale non-significant meta-analysis of four studies assessing degree of handedness was possible. When it comes to the theory of McManus (2002), the claim that left-handedness is cognitively advantageous was not supported by the present findings. Our findings do support the idea, though, that a percentage of left-handers in groups of lower cognitive ability can be attributed to «pathological» lefthandedness (Satz, 1972; Satz et al., 1985). In the meta-analyses investigating the handedness prevalence in ID individuals compared to TD individuals, the control TD groups were matched to the ID groups in terms of chronological age and

not mental age. This was necessary as only one study included a mental age control group (Porac et al., 1980). Moreover, none of the studies used adult participants. In only four studies did the participant age ranged above 18 years, but the majority of participants were again under 18 years of age (Barry and James, 1978; Pipe and Beale, 1983; Porac et al., 1980; Searleman et al., 1988). Thus, we were unable to disentangle whether the elevated levels of atypical handedness in ID individuals represents a true difference or a temporary lag on a developmental trajectory that reduces or disappears later in life. Studies investigated the question of handedness prevalence in adult ID participants are needed to address this question. An important distinction in handedness research is the distinction between hand skill and hand preference. Only studies that assessed hand preference were included in the present analyses, because the number of studies having measured hand skill were limited to two (Annett, 1993a,b; Tomprou, 2013). Preference has been shown to correlate with hand skill, albeit not perfectly (0.6–0.7; Todor and Doane, 1977). This imperfect correlation can be explained by two factors (Papadatou-Pastou et al., 2008): first, the difference in distributions between preference and skill measures (generally negatively skewed and normal, respectively) and, second, the relatively low levels of reliability of measures of relative hand skill (Hiscock and Chapieski, 2004). Nevertheless, this moderate to good correlation allows for the results of the meta-analysis to be generalizable to a certain extent to intellectual ability and hand skill. Still, we might have expected somewhat different findings if hand skill had been measured, as hand skill seems to be more sensitive in revealing handedness-intelligence correlations in general population samples (Corballis et al., 2008; Johnston et al., 2009; Peters et al., 2006) compared to hand preference (e.g., Nicholls et al., 2012). The bulk of studies included in the two main meta-analyses examined hand preference in terms of direction of hand preference and not in terms of degree of hand preference, another psychologically important distinction (McManus, 1983). This distinction is also important in terms of neural underpinnings, as it has been shown using functional magnetic resonance imaging that these aspects are independent and that they are coded separately in the brain (Dassonville et al., 1997). Moreover, it has been suggested that the strength of an individual’s hand preference may be a better indicator of underlying brain pathology and/or psychological abnormalities than the direction of hand preference (Crow et al., 1998; Nicholls et al., 2005). Yet, degree of hand preference was

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only reported in four studies comparing ID to TD individuals and including 192 participants in total. These studies were analyzed in a separate meta-analysis, but the results of this analysis did not reach significance, possibly due to low statistical power. Looking at the findings of each individual study, though, it should be noted that for all four studies the 95% CI around the standard difference of the means included zero. Thus, the limited data we have to date point to the direction of no difference in the degree of handedness between ID and TD individuals. It is hereby suggested that future studies examining the relationship between handedness and cognitive abilities report on the degree of handedness in addition to its direction. Perhaps an indication that ID individuals are more inclined toward mixed or inconclusive handedness rather than lefthandedness compared to TD individuals is the finding that the left-handedness comparison of ID to TD individuals lost significance when the largest study (Gordon, 1921) was excluded. This was not the case for the non-right-handedness comparison. Moreover, the OR of the left-handedness comparison was smaller (1.98; 1.73 without Gordon, 1921) compared to the non-right-handedness comparison (2.66; 2.60 without Gordon, 1921). Thus, when it comes to left-handedness prevalence ID individuals seem to have smaller differences (if any, as shown by the lack of statistical significance after the exclusion of Gordon, 1921) from TD individuals compared to non-right-handedness where the differences are more robust. 5. Conclusions The meta-analyses presented here show that elevated levels of atypical handedness are to be found in ID compared to TD individuals, while this is not the case for IG compared to TD individuals. The data sets used were further shown not to be susceptible to ascertainment bias, thus our findings are not due to preferential reporting of positive findings. Even though large heterogeneity was found amongst the studies comparing ID and TD individuals with regards to their handedness prevalence, the four moderator variables investigated (year of publication, classification of handedness, sex ratio, and intelligence instrument) were not found to exert significant moderating effects, possibly due to the low power of meta-analytic studies when it comes to examining moderating effects. Elevated levels of atypical handedness in ID individuals could possibly be attributed to cognitive crowding due to a re-organization in the typical pattern of cerebral laterality for language and hand control to the right and/or to pathological factors. References 3 Annett, M., 1967. The binomial distribution of right, mixed and left handedness. Q. J. Exp. Psychol. 19 (4), 327–333. Annett, M., 1974. Handedness in the children of two left-handed parents. Br. J. Psychol. 65 (1), 129–131. Annett, M., 1985. Left, Right, Hand and Brain: The Right Shift Theory. Lawrence Erlbaum Associates, London. Annett, M., 1992. Spatial ability in subgroups of left- and right-handers. Br. J. Psychol. 83 (4), 493–515. Annett, M., 1993a. Handedness and educational success: the hypothesis of a genetic balanced polymorphism with heterozygote advantage for laterality and ability. Br. J. Dev. Psychol. 11 (4), 359–370. *Annett, M., 1993b. The disadvantages of dextrality for intelligence-corrected findings. Br. J. Dev. Psychol. 84 (4), 511–516. Annett, M., 1995. The right shift theory of a genetically balanced polymorphism for cerebral dominance and cognitive processing. Curr. Psychol. Cogn. 14, 427–480. Annett, M., 2002. Handedness and Brain Asymmetry: The Right Shift Theory. Psychology Press, East Sussex.

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