Mental playmates: Siblings, executive functioning and theory of mind

Transcription

1 733 British Journal of Developmental Psychology (2006), 24, q 2006 The British Psychological Society The British Psychological Society Mental playmates: Siblings, executive functioning and theory of mind Anna McAlister and Candida C. Peterson* University of Queensland, Australia This study assessed the theory of mind (ToM) and executive functioning (EF) abilities of 124 typically developing preschool children aged 3 to 5 years in relation to whether or not they had a child-aged sibling (i.e. a child aged 1 to 12 years) at home with whom to play and converse. On a ToM battery that included tests of false belief, appearancereality (AR) and pretend representation, children who had at least 1 child-aged sibling scored significantly higher than both only children and those whose only siblings were infants or adults. The numbers of child-aged siblings in preschoolers families positively predicted their scores on both a ToM battery (4 tasks) and an EF battery (2 tasks), and these associations remained significant with language ability partialled out. Results of a hierarchical multiple regression analysis revealed that independent contributions to individual differences in ToM were made by language ability, EF skill and having a childaged sibling. However, even though some conditions for mediation were met, there was no statistically reliable evidence that EF skills mediated the advantage of presence of child-aged siblings for ToM performance. While consistent with the theory that distinctively childish interaction among siblings accelerates the growth of both ToM and EF capacities, alternative evidence and alternative theoretical interpretations for the findings were also considered. When preschool children develop a theory of mind (ToM), they gain a fundamental understanding of human psychology that allows them to interpret people s behaviour in terms of mental states such as intention, desire and belief. Even more impressively, ToM includes awareness that people may act on the world in terms of beliefs that are false, and may be guided by mistaken perceptions, erroneous memories or idiosyncratic desires and emotions. Consequently, once children acquire a ToM, more effective communication, interaction and social understanding become possible, and opportunities to engage with siblings and peers in conflict resolution and games of strategy or pretend are likely to increase (Astington, 2003; Cutting & Dunn, 1999), possibly accompanied by heightened sensitivity to criticism (Dunn, 1995) or new capacities for antisocial deception and Machiavellian manipulation (Repacholi, Slaughter, Pritchard, & Gibbs, 2003). * Correspondence should be addressed to Professor C.C. Peterson, School of Psychology, University of Queensland, Brisbane, Queensland 4072, Australia ( [email protected]). DOI: / X70094

2 734 Anna McAlister and Candida C. Peterson Empirically, the child s level of ToM understanding is typically measured using experimental tests of awareness of mental representation involving concepts such as false belief, appearance-reality (AR), desire reasoning, pretending or thinking (see Flavell, 1999, for a review of many of these tasks). The most widely used litmus test for ToM is the inferential false belief paradigm. In this procedure, participating children know the true states of affairs, but must predict the behaviours of protagonists (humans, puppets or storybook figures) who do not, but instead hold false beliefs about them. Accurate inference that the protagonist will act upon the faulty belief provides strong evidence that the child possesses a representational ToM, operationally defined as awareness of subjective mental states and their mediating role in human behaviour. Wellman, Cross, and Watson (2001) conducted a meta-analysis of 178 studies of preschoolers performance on false belief tasks like these. The results showed that the procedure is methodologically robust and reliable, and that, consistently across diverse sample populations, typically developing children s performance displays a dramatic developmental shift between the ages of 2 and 5 years. Wellman et al. found that a majority of 3-year-olds performed at chance, or below, on false belief tasks whereas, by age 5, success was so widespread across all task variations as to suggest that understanding of belief, and, relatedly, understanding of mind, exhibit genuine conceptual change in the preschool years (p. 655). Equally importantly, clear individual differences during this critical acquisition period were revealed, with some preschoolers consistently succeeding on false belief tests many months ahead of others. The focus of the present study is on these individual differences in typically developing children s rates of ToM development between age 3 and age 5. Several different explanations have been put forward to account for this variability in developmental timing. As Slaughter and Repacholi (2003) noted, research striving to explain individual differences in false belief performance falls into three broad groupings, namely: (1) studies of early experiences, and background factors in children s social-cultural and family environments, that might predict faster or slower development; (2) studies of neuroanatomical predictors and/or of links between ToM and basic cognitive skills (e.g. executive functioning); and (3) studies addressing social outcome measures, such as peer popularity and school adjustment. Siblings and ToM Notable among studies of the first type have been those that have explored the influence of the presence of siblings on ToM development. Sparked in part by naturalistic longitudinal observations (e.g. Brown, Donelan-McCall, & Dunn, 1996; Dunn, 1994, 1995) of links between children s more advanced performance on laboratory false belief tests and their frequent talk about cognitive and emotional states during family conversations with siblings and parents, the child s sibling constellation has been examined as a possible predictor variable. In one groundbreaking study, Perner, Ruffman, and Leekam (1994) tested 76 children aged 3 to 5 years on a narrative false belief test and discovered a linear improvement in performance with increasing family size. Only children did worse than their peers of the same age who had one sibling, younger and older siblings were equally beneficial, and children with two or more siblings significantly outperformed those who had only one. Perner et al. concluded, the benefit children get from interacting with two siblings rather than none is worth about as much as 1 year of experience, which at this young age is considerable (p. 1230).

3 Mental playmates: Siblings, executive functioning and theory of mind 735 Subsequently, Ruffman, Perner, Naito, Parkin, and Clements (1998) undertook a larger study that subsumed the original sample tested by Perner et al. (1994) and found that the child s number of older siblings predicted false belief scores over and above chronological age, but that once age and older siblings were considered, the presence or numbers of younger siblings exerted no additional benefits. Using a similar method, Lewis, Freeman, Kryiakidou, Maridaki-Kassotaki, and Berridge (1996) similarly concluded that older siblings were more beneficial than younger siblings to ToM development. However, when Jenkins and Astington (1996) controlled statistically for both chronological age and verbal ability, they found that preschoolers with at least one younger or older sibling significantly outperformed only children on a false belief battery, and that family size, but not birth order, was responsible for the sibling effect, so that [i]t is the number of siblings that the child has that is important for the development of false belief rather than whether these siblings are older or younger and how far distant in age they are (p. 75). As Peterson (2000) pointed out, one possible reason for the discrepancies among these studies conclusions, as well as for the disagreement between the results of all these studies showing a sibling effect and others that have not found it (e.g. Cutting & Dunn, 1999), may have been the failure of many previous studies to consider the sibling s age. Within certain age limits, sibling interaction is conceivably capable of producing distinctive kinds of social experiences that could help to accelerate a child s ToM development. However, outside these age boundaries, sibling interaction may not be so distinctive or so influential. For example, until they acquire language and social awareness, infants are probably too young to effectively play or communicate with an older brother or sister. Thus, a preschooler whose only sibling is a an infant less than 12 months old (an increasingly common situation in middle-class families today, given contemporary trends for small, widely spaced, sibling constellations) might not have optimal access to the kinds of sibling interactions that could most effectively stimulate an awareness of others minds, even though observation of parent-infant interaction might supply some clues. At the other extreme, a sibling who is an adolescent or young adult is unlikely to play or interact with the preschooler in a way that differs distinctively from a parent s. Thus, having a sibling might be most beneficial for ToM growth when child siblings are the right age to want to share together in pretend play and other childish games, and are close enough in age to be able to argue, converse, tease, joke, empathize, compete, cooperate and negotiate with on another on a relatively equal footing (Piaget, 1965). Empirically, it is unfortunate that many sibling ToM studies, including those listed above, have failed to supply full details, or conduct analyses, of siblings ages in relation to these likely optimal interaction boundaries. Thus, the possibility that samples in some studies contained mostly infant siblings too young to interact meaningfully, and/or many older siblings who functioned as adults, cannot be ruled out. Discrepancies among studies findings could conceivably be explicable on this basis. (For example, a study of participants who mostly had infant siblings could produce the result that numbers of younger siblings are of no ToM benefit, while the opposite conclusion could be reached if all the younger siblings in a particular study happened to be toddlers or young preschoolers). To test this possibility, Peterson (2000) administered a battery of false belief tests to a sample of 265 Australian children aged 3 to 5 years whose parents had supplied full information on the ages of all siblings in the participating household. As had been predicted by the interaction-opportunity hypothesis, children whose only siblings were

4 736 Anna McAlister and Candida C. Peterson infants under the age of 12 months were found to display no sibling ToM advantage, performing like only children on false belief tests. The same was true of participants whose only siblings were teenagers or adults. However, those preschoolers in Peterson s sample who had at least one younger or older sibling in the childhood age range scored significantly higher on false belief tests than both only children and those who only had infant or adult siblings (dubbed pseudo-singletons). Middle-born children who shared a household with several child siblings, including those younger and older than themselves, did best of all, possibly owing to the benefits of the greater variety of childish interactional opportunities that siblings of mixed ages could supply. Peterson concluded, [t]he opportunity to play, converse and disagree in distinctively childish ways with brothers and sisters provides unique insights into the workings of the human mind (p. 435). While this conclusion seems plausible in terms of interaction opportunity, it is important to test the replicability of the sibling age boundary effect, and how well it generalizes to other measures of mentalistic understanding besides false belief (these were the only ToM tests Peterson (2000) used). As subsidiary aims, these questions will be addressed in the present study. Executive functioning, ToM and siblings Slaughter and Repacholi s (2003) second type of account for individual differences in preschoolers rates of ToM development centres on differences in cognitive ability, especially executive functions (EF). Executive capacities are domain-general problemsolving tools. Children with sound EF skills are able to exert self-control and to strategically inhibit prepotent, but incorrect, responses (Perner & Lang, 1999). They are also able to think flexibly and planfully about varied kinds of problems while switching rapidly among different modes of processing information (Carlson & Moses, 2001). Specific executive skills that are often experimentally measured include forward planning, mental flexibility, working memory and response inhibition. Preschoolers scores on ToM tasks have been shown to correlate with scores on tests of these EF abilities (Carlson & Moses 2001; Carlson, Moses, & Hix, 1998; Hughes, 1998). For example, Carlson and Moses found that, even after partialling out children s language ability and age, scores on an EF battery of inhibitory control tasks remained strongly correlated with scores on a false belief battery. Theoretically, some accounts of ToM development (e.g. Hughes, 1998) postulate that EF abilities are necessary prerequisites for the basic conceptual understanding of mental states that defines ToM and/or for the execution of accurate responses to ToM tests. One possibility is that children with underdeveloped EF skills may not be capable of conceptualizing representational mental states like false belief. Then again, their implicit conceptual understanding may fail to produce success on standard tasks owing to the executive demands for memory, attention, inference and response inhibition that are bound up with the false belief testing procedure. Alternatively, ToM abilities are sometimes thought of as causally and developmentally primary to EF skills (e.g. Perner & Lang, 1999), so that an initial capacity to understand ToM concepts like false belief may underpin the later development of EF abilities. For example, children may need to reflect on the mind and understand its links with overt behaviour before they can deliberately employ the cognitive strategies identified as executive functions. In yet another intriguing theoretical slant on the relationship between EF and ToM, Cole and Mitchell (2000) postulated that the presence of siblings might accelerate children s development of EF abilities, with these EF advances, in their turn, accelerating

5 Mental playmates: Siblings, executive functioning and theory of mind 737 the development of ToM. As they suggested, [i]t might be that having siblings primarily promotes executive control (p. 281) so that any effect of siblings on ToM performance is mediated through individual differences in EF. As Cole and Mitchell noted, the suggestion that siblings could be beneficial to preschoolers EF performance might initially seem unlikely. As they point out, tests of executive functions often correlate with tests of intelligence (p. 280). Yet, Zajonc, Markus, and Markus (1979) found negative correlations between intelligence test (IQ) scores and the total number of siblings in the child s family, and Glass, Neulinger, and Brim (1974) found that only children and eldest offspring in widely spaced families (who had grown up as only children for a significant time) scored higher than laterborns on tests of reading ability and scholastic achievement as well as IQ. However, on plausible theoretical grounds, Cole and Mitchell made the opposite prediction for EF in preschoolers. Specifically, they argued that young children with siblings might develop EF skills rapidly through play and sibling interaction that could cultivate EF skills for remembering, planning strategically and inhibiting prepotent incorrect responses in a manner not available to the only child. For example, siblings may react more negatively than parents to one another s failures to share, plan, attend or control their impulses. Furthermore, while playing strategic games, negotiating over conflicts, delaying gratification or taking turns, siblings might observe, approve, teach and copy one another s EF skills. However, when they conducted two experiments that were designed to test whether the sibling ToM advantage reflected a more basic beneficial effect of siblings upon EF abilities, Cole and Mitchell (2000) obtained null results. In a sample of predominantly low-income families, they found mostly non-significant correlations between preschoolers EF scores and their numbers of siblings. Nor was there any evidence of a sibling ToM advantage, since non-significant correlations emerged in both experiments between all tasks in their ToM battery and each aspect of sibling numbers (total, older or younger) that they scored. In fact, the only exception to this predominant pattern of non-significant findings was a weakly significant correlation (r ¼ :24) in the second experiment between sibling numbers and one of their four EF tests. Yet, this same correlation was far from significant in the first experiment (r ¼ :04) and it likewise fell to non-significance in the second experiment once differences in children s verbal abilities were statistically controlled. Cole and Mitchell speculated that their sample s socio-economic disadvantage may have contributed to these null results. In fact, degree of socio-economic disadvantage negatively predicted false belief scores in their second experiment even after age and verbal ability were partialled out. It makes some sense that explanatory, mentalistic discourse among siblings, and between siblings and parents, might be less frequent in lower-income families than in middle-class homes. Thus, the null results might have reflected a confound between family size and economic status for the particular groups of children Cole and Mitchell had tested. Yet, when Hughes and Ensor (2005) studied similar questions among 2-year-olds who were also predominantly from lower-income families, they did observe a sibling influence of a different kind. As in Cole and Mitchell s study, there were no significant correlations between the toddler s total number of siblings and his or her scores on ageappropriate batteries of ToM and EF tasks. However, the quality of the children s relationships with their siblings (a variable reflecting maternal reports of their frequent pretend play, frequent affectionate communication, etc.) was a statistically significant predictor both of toddlers ToM performance and of their EF scores. Furthermore, the link between ToM and sibling relationship quality remained significant even after

6 738 Anna McAlister and Candida C. Peterson the influences of age, EF, linguistic maturity and parent-child relationship quality were statistically controlled. Since Hughes and Ensor did not report the ages of their participants siblings, it is unclear to what extent sibling interaction opportunity, as a function of age, might have contributed to the findings. Infant siblings may have been too young, and adolescent siblings too old, to engage in the kinds of childish conversations, pretend play and so on, that would not only produce high relationship quality scores but also, according to Peterson (2000), promote the rapid development of ToM. Aims of the present study The main aim of the present study was therefore to further examine the extent to which the presence or absence of child siblings, and sibling numbers and birth order, might be linked with individual differences in preschooler s performance on ToM and EF tasks. By including measures of sibling age, we were able to indirectly test the sibling-interactionopportunity hypothesis (Peterson, 2000); namely, that siblings should be children to offer optimal ToM benefit. If it is the opportunity for distinctively childish play and conversation with siblings that is important, then we predict that children with at least one child sibling (aged 1 to 12 years) will outperform only children and pseudosingletons (i.e. those with only infant or adult siblings) on batteries of standard ToM and EF tasks. In addition, by examining a larger sample of children than in Cole and Mitchell s (2000) study, as well as a group more economically advantaged than theirs, and by including additional ToM tests of AR and pretend representation (PR) as well as false belief, subsidiary aims of the study included more extensive testing of two plausible hypotheses put forward by Cole and Mitchell, namely: (1) that a more affluent sample of preschoolers than in Cole and Mitchell s study would reveal clearer links between the three target variables (child siblings, EF and ToM), and (2) if so, that evidence of superior ToM performance by participants with child siblings would be mediated through a more basic child-sibling influence upon EF. Method Participants After the exclusion of 21 children (3 for reports of a suspected diagnosis of developmental delay, 3 for a parental report of a primary family language other than English, 6 owing to parents failure to specify the sibling composition of their family and 9 for failing a control question on a false belief test; see below), the sample consisted of 124 Australian children (71 boys and 53 girls) who ranged in age from 3; 3 5; 9 with a mean of 4; 3. The children were attending preschools that served predominantly middleclass neighbourhoods, all had English as a sole or primary language and none had a known or suspected serious intellectual, developmental or sensory impairment. In addition to predominantly middle-class residence, their parents were well educated (averaging some university study ; see Table 2). In terms of sibling composition, 18 children in the sample (14%) were only children who had no brothers or sisters of any age. An additional 5 children resembled only children in terms of their family conversational environment (Peterson, 2000) in that their sole sibling was either a very young infant (2 children) or an adult (3 children). For the statistical analyses, these groups (singletons plus pseudo-singletons) were merged together as Group 1, defined as participants with no child siblings. They comprised 19% of the sample. Group 2 consisted of 49 children (40% of the sample) who had at least

7 Mental playmates: Siblings, executive functioning and theory of mind 739 one older child sibling (under age 13) but no younger child siblings. Of these, 40 children were the youngest in their families while an additional 9 also had an infant sibling under 12 months of age. Group 3 consisted of 21 children (17% of the sample) who had both older and younger child siblings and Group 4 consisted of 31 children (25% of the sample) who had younger child sibling(s) but no older ones. Tasks and scoring The children completed measures of ToM (four tasks) and EF (two tasks) plus a language measure. These tests are described below. Theory of mind (ToM) The ToM battery was chosen to include widely used tests of representational understanding that, like those used by Hughes and Ensor (2005) varied sufficiently in difficulty level to be able to pick up early signs of ToM awareness while also emphasizing the standard litmus ToM measures like false belief. The ToM tests were of three different kinds; namely, false belief, AR and pretend representation. All three of these were alike in requiring children to understand mental representation together with the fact that people s inner psychological states can misrepresent objective reality. Specifically, to pass the false belief tests, children had to predict actions of protagonists whose beliefs misrepresented the real locations of objects or contents of containers. To pass AR, they needed to understand that an object s perceptual appearance can misrepresent its true nature and function, and to pass the pretend tasks, they needed to understand the distinction between an object s real properties and its alternative representational identity during pretending. False belief. There were two false belief tests: (a) a standard unseen displacement task and (b) a standard misleading container task. For (a), we used Baron-Cohen, Leslie, and Frith s (1985) Sally-Ann task, presented and scored exactly as described by the original authors except that a boy replaced the second girl doll (Peterson & Siegal, 1999). In brief, each trial began with the girl doll placing a marble in a covered basket and going away. The boy doll then moved the marble either to a covered box (Trial 1) or to the tester s pocket (Trial 2). The girl returned wanting her marble. Children were asked the test question, Where will the girl look first for her marble? and then two control questions ( Where is the marble really? and Where did the girl put the marble in the beginning? ). When children passed both control questions on both trials, it was clear that they had enough language to understand the tasks and test questions and that they had attended to, and remembered, the critical information on which a valid test question response could be based. Thus, we required that all children pass these control questions in order to be included in our sample. (In the rare instance where a child we had begun to test failed one of these four control questions, he or she was not tested any further and was not included in our sample or analyses). Each correct test question response earned 1 point to a maximum total of 2 for the task. The misleading container false belief test was borrowed from Gopnik and Astington (1988) and involved a familiar Band-Aid box that actually held pencils. The child was initially shown the closed box and was asked what was in it. After naming the expected contents (which all did), the pencils were displayed and the box was resealed. The control question ( What is really in the box ) was given, followed by a test question about other s belief: [name of class mate] is coming next. He [she] hasn t seen inside this box before. When I show it to him [her] all closed up like this, what will he [she] say is in it? Next was a representational change test question about own false belief:

8 740 Anna McAlister and Candida C. Peterson When I first showed you this box, before you looked inside, what did you think was in it? followed by a control question: What is really in it? All children passed both control questions. Correct test question responses (Band-Aids) earned 1 point to a possible total of 2 for the task. Children s scores on the two false belief tasks were found to be statistically significantly correlated, rð122þ ¼:50, p, :001. Consequently, a total false belief (TFB) score was created by summing scores on the changed location and container tasks. This total could (and did) range from 0 4. Appearance-reality (AR). This test was based on Flavell s (1986) procedure. There were two trials. The first began with the display of a realistic wax apple that was really a candle. On being asked what it was, all children said apple. Then the tester demonstrated the real identity by lighting the candle and asked the two test questions in counterbalanced order: What is this really and truly? (reality test) and When you look at this with your eyes right now, what does it look like, [pause] does it look like a [candle] or like [an apple]? (appearance test). Accurate understanding of the appearance reality distinction requires that the child distinguish the two identities, so that a child who gets only one test question right (typically by giving the same answer, e.g. apple, to both test questions) has demonstrated no greater understanding one who gets both wrong. Indeed, the latter child (who has typically distinguished appearance from real identity while transposing their labels) could be argued to have somewhat greater understanding than the former. Consequently, in order to pass, children had to answer both test questions correctly and, if so, earned 1 point. Other response patterns earned 0. A second AR trial using a realistic-looking cloth flower that was actually a ballpoint pen was conducted using identical procedures, questions and scoring to the apple-candle test. Scores on the apple-candle and flower-pen versions were significantly correlated, rð122þ ¼:34, p, :01. Consequently, a total AR score ranging from 0 2 was created by summing scores on both of them. Pretend representation (PR). The third component of the ToM battery assessed children s awareness of the representational nature of pretending and of the contrast between an object s real, physical identity and its mental representation as something else during pretence. There were two trials. Each began with the child being shown a familiar, real object (banana or potato, counterbalanced) and being asked to pretend it was, respectively, a phone or a bar of soap. After minimal prompting (e.g. Your phone s ringing ), all children successfully engaged in actions that unequivocally showed they understood and accepted the pretend stipulation (e.g. by holding the banana to their ear, or rubbing the potato over both hands). After a short period of pretend play, the tester said Okay, we ve finished our game now. We have stopped pretending. Then, the tester pointed at the banana or potato and asked two test questions: (1) What is this really? and (2) What did we pretend this was? Two additional objects were then placed on the table alongside the banana or potato. One was always a real version of the pretend item (e.g. a real mobile phone) and the other was an irrelevant decoy (e.g. a tennis ball). The last two test questions for each trial were: (3) Which one did we pretend was a [telephone/soap]? and (4) Which one is really a [phone/soap]? Thus, for each PR trial, there were four test questions, using different syntax, two about the real, and two about the apparent, identity of the object. For each correct test question response, 1 point was awarded, to a maximum of 4 for each task. Scores on the two tasks were statistically significantly correlated, rð122þ ¼:31, p, :01, so children s scores on the two tasks were summed, then halved to give a total out of 4, similar to the false belief task.

9 Mental playmates: Siblings, executive functioning and theory of mind 741 Total ToM score The children s total FB scores were significantly correlated with their total AR scores, rð122þ ¼:29, p, :01, and with their total PR scores, rð122þ ¼:43, p, :01. The PR total was also significantly correlated with the AR total, rð122þ ¼:30, p, :01. Consequently, given these significant associations and in order to create a total ToM score that would given equal weighting to each of these three frequently assessed aspects of representational understanding, a proportion score for each component was first computed by dividing the FB and PR totals by 4, and the AR total by 2. These decimal totals were then averaged into a ToM composite for each child that could range from 0 1. Cronbach s test was run on this ToM composite and produced an alpha reliability coefficient of.65, indicating acceptable internal consistency for the ToM composite as a summary score. Executive functioning (EF) Two tests of EF were borrowed from Cole and Mitchell (2000). The first was a route navigation (RN) test assessing EF capacities for forward planning, attention to relevant features, resistance to distraction, and response inhibition. The task was presented and scored exactly as described by Cole and Mitchell. There were four trials. On each trial, the child was shown a symmetrical maze printed on coloured card with a picture of a cartoon character at the entrance (middle-bottom) and a picture of a desired goal (e.g. a gift-wrapped present) at the middle-top (exit). The paths on the right and left side of each maze were both of equal length and both had 10 bends. One path (left or right, counterbalanced over trials) was unobstructed, while the other was blocked at three points by brightly coloured stickers. Children were told that the route with the stickers was the hard way for the character to go because each sticker represented an obstacle that had to be climbed over. The route without stickers was described as the easy way because it had no blockages. Children were then asked to draw with their finger the easy way for the character to go to reach the goal. On the first trial only, corrective feedback was given to any child who made an error, in which case the trial was repeated. After this, without correction, children earned 1 point for tracing the unobstructed path, otherwise 0, to a maximum possible score of 4 for the task. This score was then converted to a proportion of trials correct by dividing by 4. The second EF test was Cole and Mitchell s (2000) adaptation of Reed, Pien, and Rothbart s (1984) resisting instructions (RI) test. We followed Cole and Mitchell s procedure except that a bear and lion replaced their squirrel and badger for the sake of familiarity to local children. This task assesses EF skills for response inhibition, task switching and adherence to directions. In brief, over 12 trials, the bear and lion puppets took turns asking the child to perform various movements (e.g. touch your nose ). The hand on which the tester wore bear and lion puppets was counterbalanced across participants. Before the test began, children were told that the lion was naughty and should not be obeyed. Instead, they should keep their hands flat on the table whenever the lion instructed them to do something. However, the bear was good so they should always do exactly what the bear told them to. Initially, and again at the start of every fourth trial, the tester asked the child to point to the puppet who should be obeyed and to the one who should be ignored. In randomized order, the bear gave the command on 6 trials, and the lion on the remaining 6 trials. The lion s trials were test trials and the bear s were filler trials and were not scored. Children earned 1 point for each trial in which they ignored the lion and kept their hands perfectly still. Thus, the maximum

10 742 Anna McAlister and Candida C. Peterson possible total was 6 for the task as a whole. To yield a total similar to the RN, this score was converted to a proportion by dividing by 6. Scores on the RN and RI tasks were statistically significantly correlated, rð122þ ¼:33, p, :01, so children s scores on the two tasks were averaged to yield a composite EF total score. Language ability The Peabody Picture Vocabulary Test (PPVT; Dunn & Dunn, 1981) was administered and scored according to standard instructions to give three estimates for each child. These were: (1) the raw score (a measure of receptive vocabulary size), (2) a verbal mental age estimate (VMA; or language maturity relative to children in Dunn & Dunn s 1981 American normative sample) and (3) a standard score (a deviation-type norm, or Verbal IQ equivalent). Parent education The parent who supplied information on sibling composition, home language and disability diagnosis also completed a brief scale of educational attainment for self and spouse by selecting, in response to the question, What is the highest level of education presently achieved?, a response option from the following: (a) Year 10 only (the minimum legal school leaving age), scored 1; (b) high school or technical school completion (12 years of schooling), scored 2; (c) some university studies but no degree, scored 3; (d) university Bachelor s degree completed, scored 4; (e) university Masters or PhD degree completed, scored 5. Father s and mother s scores were averaged, for the 119 intact families in the sample, to produce a parent mean score that could range from 1 5. For the five single mothers without a co-resident partner, the mother s score was used without averaging. Testing procedure Each child was tested individually in a quiet area of the preschool in two sessions, separated by about one week. Each session include a single member of the FB, AR, PR and EF task pairs. The specific task of each type that was given first or second was counterbalanced. The PPVT was always given in the second testing session. Results Descriptive information and preliminary analyses Table 1 shows the children s mean scores on the ToM and EF measures. In order to confirm that children in this sample were developing on the typical timetable that has Table 1. Children s performance on ToM and EF tasks ToM battery EF battery False belief (FB) Pretend representation (PR) Appearance reality (AR) Route navigation (RN) Resisting instructions (RI) Possible score Mean score (SD) 2.54 (1.44) 3.52 (0.64).78 (0.82) 3.19 (1.05) 3.25 (2.69) Range

11 Mental playmates: Siblings, executive functioning and theory of mind 743 been established in much previous research (see Wellman et al., 2001, for a review), age trends for our sample s ToM scores were examined. As expected based on these earlier studies, ToM composite scores for the children in this sample were significantly correlated with age, rð122þ ¼:31, p, :01, two-tailed. Furthermore, when the sample was subdivided into three age groupings of (1) 3-year-olds (N ¼ 42; mean age ¼ 3.72), (2) young fours (N ¼ 43; mean age ¼ 4.23) and (3) older preschoolers (N ¼ 39; mean age ¼ 4.87), a significant age difference in ToM scores emerged, Fð2; 127Þ ¼4:72, p, :05. To locate the basis for this difference, single-df planned comparisons were used (Keppel, 1982). The 3-year-olds, with a mean proportion of 0.67 ToM test trials correct, were found to be significantly less advanced than the young 4-year-olds who, with a mean of 0.78, Fð1; 122Þ ¼8:11, p, :01, did as well as the older preschoolers, whose mean was.77, Fð1; 122Þ, 1, ns. There were no statistically significant differences among the three age groups in level of parent education, F, 1, child gender distribution, x 2 ð2; 124Þ ¼2:33, ns, or PPVT standard scores, F, 1. Thus, the age effect was an uncomplicated one, and it supported previous findings of a rapid surge in ToM development after the fourth birthday (Wellman et al., 2001). Consequently, in contrast to Cole and Mitchell s (2000) socio-economically disadvantaged sample, there was no reason to anticipate lack of sibling influences on ToM for children in the present sample owing to delayed mentalistic understanding. As a further preliminary analysis, we examined the comparability of the four sibling constellation groups on variables that could potentially confound the assessment of sibling effects on ToM and EF. Table 2 gives shows the groups mean scores on these variables. Results of the preliminary analyses revealed that there were no statistically significant differences among the sibling constellation groups in age, Fð3; 120Þ ¼:53, ns, in gender balance, x 2 ð3; 124Þ ¼6:27, p ¼ :10, in mean levels of parents education, Fð3; 120Þ ¼1:49, ns, or in language maturity (VMA), Fð1; 122Þ, 1, ns. Thus, neither the contrast between only children and those with a child sibling, nor the participant s status as a youngest, middle or eldest child sibling, displayed any consistent associations with any of these other potential predictors of ToM or EF, consequently simplifying the main statistical analyses, and indicating that any differences we observe in ToM or EF abilities as a function of sibling status will not be explicable as an indirect consequences of these other variables. Child siblings and ToM development To test the set of hypotheses relating to the influence of child siblings on individual differences in rates of ToM development, an analytic set of single-df planned comparisons was conducted, in the manner recommended by Keppel (1982) directly on [the] data without reference to the significance or non-significance of the omnibus F test (p. 106) and instead of the overall F test (p. 104, emphasis in original text). The first hypothesis, derived from Peterson (2000), was that the presence of at least one child sibling in the family would be linked with significantly more advanced ToM than for Group 1 children without siblings at home. This was tested with a planned comparison contrasting Group 1 s ToM composite score with that of the combined remainder of the sample. The result was statistically significant, Fð1; 120Þ ¼6:08, p, :025, indicating that the children without child siblings in Group 1 were significantly less advanced in ToM development than those who had at least one brother or sister aged 1 to 12 years. Next, to test the hypothesis derived from Ruffman et al. (1998) that older child siblings might facilitate ToM growth whereas younger siblings would not, a planned comparison was conducted between Groups 2 and 3 combined

12 744 Anna McAlister and Candida C. Peterson (both of whom had older child siblings) versus Group 1 plus Group 4 (without older child siblings). The result was not statistically significant, Fð1; 120Þ ¼2:72, ns. A different way of testing the same hypothesis, using the strategy employed by Jenkins and Astington (1996), is to compare children who have older siblings only with those whose only siblings are younger, excluding middle-borns. This planned comparison, between Group 2 and Group 4, was similarly non-significant, F, 1, ns. Finally, in order to test the hypothesis from Perner et al. (1994) that children with two or more child siblings develop ToM faster than those with only one child sibling, a planned comparison was conducted between Group 3 (who had a mean of 2.10 child siblings) and Groups 2 and 4 combined (who had a mean of 1.05 child siblings). This comparison was likewise nonsignificant, F, 1, ns. In other words, children in the present sample who had no child siblings scored significantly below those who had a child sibling on our composite ToM measure. These findings support Peterson s (2000) for children from a similar population. In particular, using a different ToM battery that included additional AR and PR tasks, our results replicated Peterson s findings (1) that better false belief performance was associated with having at least one child sibling, and (2) that the sibling s being older, rather than younger, than the participant was not additionally beneficial. Siblings and EF Table 2 shows the mean EF composite scores for children in each of the sibling constellation groups. The possibility that having a child sibling would influence EF performance was tested using single-df planned comparisons (Keppel, 1982). For the first comparison, participants without child siblings (Group 1) were contrasted with the combined remainder of the sample (Groups 2, 3 and 4) who had at least one child sibling aged 1 to 12 years. This contrast was statistically significant, Fð1; 120Þ ¼4:12, p, :05. In other words, participants with a child sibling did better on EF tasks than only children plus those whose only sibling was an infant or an adult. The second planned comparison evaluated influences of the presence of an older sibling by contrasting the EF scores of Groups 1 and 4 with those for Groups 2 and 3 Table 2. Performance by children in different sibling status groups on theory of mind, executive functioning and language measures, along with other sample characteristics Participant s sibling situation Group 1. No child siblings 2. Has older child sibling(s) only 3. Has younger and older child siblings 4. Has younger child sibling(s) only Group size N ¼ 23 N ¼ 49 N ¼ 21 N ¼ 31 Mean age (SD) 4.14 (0.59) 4.26 (0.59) 4.32 (0.48) 4.30 (0.48) Age range 3;3 5;2 3;3 5;9 3;6 5;5 3;3 5;4 Mean parent 3.16 (1.01) 3.12 (1.17) 3.40 (1.08) 3.65 (1.18) education (SD) Gender balance 13:10 25:24 17:4 16:15 (male: female) Mean VMA (SD) 4.47 (0.97) 4.41 (1.12) 4.63 (0.96) 4.80 (1.51) Mean ToM Total (SD) 0.66 (0.19) 0.75 (0.17) 0.78 (0.21) 0.75 (0.20) Mean EF Total (SD) 0.65 (0.27) 0.73 (0.27) 0.85 (0.18) 0.74 (0.26)

13 Mental playmates: Siblings, executive functioning and theory of mind 745 (participants with at least one older child sibling). The result approached, but did not achieve, statistical significance, Fð1; 120Þ ¼ 3:50, p ¼ :06. The alternative way of testing benefits of older versus younger siblings (by contrasting Group 2 with Group 4) was likewise non-significant, F, 1, ns. In other words, for the children in this sample, higher EF scores were linked with the presence of a child sibling, whether younger or older, rather than being confined specifically to the presence of older siblings who might serve tutoring, or other pseudo-adult, roles. Correlations of ToM and EF with sibling variables Table 3 shows the Pearson univariate correlations that emerged among key variables in this study. In line with much previous research, there were significant correlations of children s ToM and EF scores with their chronological age and verbal ability (PPVT raw scores). With verbal ability partialled out, however, the correlation between ToM and EF remained statistically significant, rð121þ ¼:37, p, :001. Within this relatively welleducated sample, parent education was not significantly related to any variable we measured. Table 3. Univariate correlations among children s scores on theory of mind, executive functioning, language measures and background variables Variable: CA #Sibs ChSib VMA ToM EF Chronological age (CA) **.31**.54** Number of child siblings (#Sibs) ** *.20* Presence of child sibling (ChSib) **.17 Verbal mental age (VMA) **.46** ToM total (ToM) ** EF total (EF) 1.00 Note. * denotes significant at p, :05 two-tailed, ** at p, :01, two-tailed In line with the ANOVA results, ToM scores were positively correlated with the presence of at least one child sibling in the participant s household (dummy coded as 0 or 1, as recommended by Tabachnick & Fidell, 1989), and this correlation remained significant even after verbal ability was partialled out, rð121þ ¼:29, p, :01. Similarly, the child s total number of child-aged siblings was correlated with ToM, and remained significant after partialling out verbal ability, rð121þ ¼:23, p, :02. Like ToM, EF scores were significantly correlated with the preschooler s number of child siblings, rð121þ ¼:20, p, :03, and this correlation remained significant with verbal ability partialled out, rð121þ ¼:23, p, :02. EF scores were also marginally correlated with the presence of at least one child sibling in the participant s household ( p ¼ :07) and this correlation rose to statistical significance after verbal ability was partialled out, rð121þ ¼:23, p, :02. Overall, then, in contrast to Cole and Mitchell s findings for lower-income children of similar age to the present sample, but partially in line with Hughes and Ensor s (2005) findings for younger lower-income children, we found that individual differences among children in our sample in both ToM and EF performance were linked with their access at home to opportunities for play and conversation with child siblings who were old enough to be genuine interactional partners, yet not too old to converse or play in distinctively childish ways.

14 746 Anna McAlister and Candida C. Peterson Furthermore, correlations of the child sibling variables with ToM and EF remained statistically significant even after differences in children s language scores were statistically controlled. Hierarchical multiple regression analyses In order to examine more closely the separate contributions of child siblings to individual differences in children s ToM and EF scores, even after accounting for the mutual dependency of both EF and ToM on age and language, we conducted a number of hierarchical multiple regression analyses. For the first set of analyses, ToM scores served as the dependent variable. Age and PPVT raw scores were entered together as control variables at Step 1 in each analysis. For the first analysis, the variable presence of a child sibling (dummy coded: 1 ¼ at least one child-aged sibling versus 0 ¼ none, as recommended by Tabachnick & Fidell, (1989)) was entered at Step 2. At the end of Step 1, with only the control variables in the equation, there was a significant effect, Mult. R ¼ :42, R 2 ¼ :18, Adj. R 2 ¼ :17, p, :001. At the end of Step 2, with the presence of a child sibling included, there was a statistically significant increment in the amount of ToM variability that was predicted, FðchangeÞ ¼8:66, p, :01. In other words, information on whether or not the preschooler had access at home to a child-aged sibling improved the prediction of differences in ToM scores beyond that afforded by chronological age and verbal ability combined. With EF scores entered at the third step, a further significant increment emerged, FðchangeÞ ¼11:52, p, :01. In other words, information on differences in the children s EF abilities further improved the prediction of differences ToM, beyond that afforded by age, language and child siblings. At the end of this final step, with all the predictor variables included, the full model was statistically significant, Mult. R ¼ :56, R 2 ¼ :32, Adj. R 2 ¼ :29, p, :01, and three independent variables were found to have statistically significant beta weights in the final equation. These were verbal ability (b ¼ 0.32, t ¼ 3:33, p, :01), the presence of a child-aged sibling (b ¼ 0:22, t ¼ 2:60, p, :02), and EF scores (b ¼ 0:33, t ¼ 3:39, p, :01). Thus, language, having a child sibling, and EF performance each made significant independent contributions to predicting individual differences in children s ToM understanding. To test whether order of entry of the EF and sibling variables made a difference to the outcome, we conducted a similar hierarchical regression with the ordering of Steps 2 and 3 transposed, so that EF scores were entered immediately after the control variables (age and language), and the presence of a child sibling was entered at the last step. The result was very similar to that with the original order of entry. The presence of a child sibling produced a statistically significant increment at the final step, even after variability due to EF, age and language had been taken into account, FðchangeÞ ¼6:74, p, :02, and the beta weights at the end of the final step were the same as in the initial regression equation, with verbal ability, EF and child siblings each making significant independent contributions. A third set of hierarchical multiple regressions predicting ToM with numbers of child siblings in place of the previous child sibling variable (presence/absence) likewise produced similar results except that, in this case, the increment was statistically significant when the number of child siblings was introduced at the second step, FðchangeÞ ¼ 5:12, p ¼ :03, but only marginally significant when number of child siblings was entered after EF at the third step, FðchangeÞ ¼2:57, p ¼ :11. In other words, comparing all these regression models, variability in children s ToM performance was clearly predicted, over and above language and age, by EF skill

15 Mental playmates: Siblings, executive functioning and theory of mind 747 and by the presence of a child sibling, and presence of at least one a child sibling was a more important independent predictor than the total number of child siblings in the household. A second set of hierarchical multiple regression analyses was conducted with EF scores as the dependent variable (DV) and ToM and numbers of child siblings as predictors. As for the ToM regressions, child age and PPVT raw scores were entered together as control variables at Step 1. The equation was significant at the end of this step, Mult. R ¼ :56, R 2 ¼ :31, Adj. R 2 ¼ :30, p, :001. When number of child siblings was entered at Step 2, there was a statistically significant increment in the variability in EF scores that was predicted, FðchangeÞ ¼5:17, p, :05, and the beta weight for number of child siblings was separately significant, b ¼ 0:18, t ¼ 2:27, p, :05. ToM scores were entered at the third step and a further significant increment emerged, FðchangeÞ ¼10:76, p, :01. At the end of this final step, with all the predictor variables included, the full model was statistically significant, Mult. R ¼ :64, R 2 ¼ :40, Adj. R 2 ¼ :38, p, :001, and two independent variables were found to have statistically significant beta weights in the final equation; namely, chronological age (b ¼ 0:40, t ¼ 4:48, p, :001), and ToM scores (b ¼ 0:28, t ¼ 3:28, p, :01). A similar result emerged when ToM scores were entered at Step 2 and number of child siblings was entered last. Thus, the number of child siblings, while significant beyond age, did not independently contribute to predicting variability in children s EF performance once both age and ToM scores were considered, suggesting that number of child siblings overlapped with ToM in predicting EF skill. Test of EF as a mediator A final set of regression analyses was conducted in order to test Cole and Mitchell s (2000) hypothesis that the effect of child siblings upon individual differences in preschoolers ToM performance might be mediated through a more basic effect of siblings upon EF, which might then drive the ToM advantage. We followed the Baron and Kenny s (1986) approach to testing mediation, using the regression design that was recommended by Holmbeck (1997). In the model we tested, with effects of the control variable (chronological age) statistically controlled, as recommended by Holmbeck, and with the number of child siblings as the independent variable (IV), EF performance as the mediator (M), and ToM scores as the DV, three out of required four conditions for mediation were met, namely: (1) the participant s number of child-aged siblings (IV) significantly predicted the mediator (EF), p, :05, (2) the participant s number of childaged siblings (IV) significantly predicted the DV (ToM), p, :05, and (3) EF performance (M) significantly predicted ToM (the DV) when it was entered together with number of child siblings in a standard regression equation, p, :01. However, the fourth condition for mediation was not met; namely, that there should be a significant reduction in the ToM variability that was predicted by the IV (child siblings) after inclusion of the mediator (EF). The results of a Sobel test indicated that this reduction was not statistically significant, Sobel ¼ 1.83, p ¼ :07. Furthermore, a Goodman I test (recommended by Baron and Kenny, 1986, as an alternative to the Sobel test) showed the same thing: Goodman I ¼ 1:78, p ¼ :07. In other words, the data failed to significantly support mediation. Even though the variables were intercorrelated, there was no statistically reliable evidence that the advantage of numbers of child siblings for ToM was channelled through EF.

16 748 Anna McAlister and Candida C. Peterson Discussion The present results for ToM performance replicated Peterson s (2000) finding of more advanced false belief understanding among preschoolers with at least one child-aged sibling at home with whom to play and converse than for only children and pseudosingletons who had no child siblings. Our results also importantly extended upon these earlier findings by showing that the child-aged sibling advantage applied also to preschoolers EF performance (in line with Cole and Mitchell s 2000 prediction but not their empirical findings) as well as to a broader range of ToM capacities for a representational understanding of pretence and perceptual appearance, in addition to false belief. Furthermore, results of hierarchical multiple regression analyses showed that sibling interaction opportunity (operationalized as having at least one child-aged sibling at home with whom to play and converse) explained additional individual variability in children s ToM scores beyond that accounted for by age and linguistic maturity. EF scores also predicted ToM performance even after the effects of language skill and age were considered, and when all four of these predictors were considered together at the final steps of hierarchical regressions, significant independent contributions to ToM variability were found to arise out of three of them, namely: (1) sibling interaction opportunity, (2) language ability and (3) EF skill. Hierarchical regression analyses were also conducted with EF scores as the dependent variable. Results of these indicated that numbers of child siblings predicted variability in preschoolers EF performance beyond that explained by chronological age and language ability alone. However, child siblings did not independently contribute to predicting variability in EF performance once both age and ToM were considered. A number of causal accounts for results like these are conceivable and, on the basis of a cross-sectional correlational design like ours, no single developmental or causal pathway can be uniquely identified. Longitudinal investigation of the same variables would be useful for this purpose. Nevertheless, purely speculatively, it is possible that the nature of the social interaction that goes on in a preschooler s family when a child sibling is present differs qualitatively from that available to a singleton or pseudosingleton whose only social companions are parents or other adults. The added social experiences available in the former instance could conceivably underpin the childsibling advantages for ToM and EF performance. For example, as Cole and Mitchell (2000) suggested, a preschooler s EF skills for processing information strategically and flexibly while inhibiting impulsive responses could be enhanced through experiences with siblings of playing competitive, cooperative and strategic games, by older siblings modelling or coaching of these skills, or via sibling disapproval when EF skills are not manifested. Similarly, as Hughes and Ensor (2005) suggested, frequent opportunities to play imaginatively with siblings, to discuss thoughts and feelings and to resolve sibling conflicts could stimulate not only the social awareness that is bound up with ToM but also the cognitive flexibility, inhibitory control, rule-consistency and problem-solving skills that are assessed by many EF measures. Alternatively, it is conceivable that other factors, and a different chain of influence, might account for the correlations we observed. For example, parents who plan their families with close sibling spacing because they want to make sure each of their children has companionship from a similar-aged brother or sister might also differ in other ways from parents with different family goals, or those who do not actively plan child-bearing. Associated differences in parental philosophies or child rearing strategies might then indirectly explain the correlations between child siblings and both EF and

17 Mental playmates: Siblings, executive functioning and theory of mind 749 ToM performance. It is also likely that parents styles of talking with their offspring about thoughts and feelings may have a bearing on how readily the children develop ToM (Brown & Dunn, 1992; Dunn, 1994; Peterson & Slaughter, 2003). The present results do not inform directly on how variations in parents conversational approaches might relate to their families sibling constellations and it would be useful for future research to study this. Interestingly, however, Hughes and Ensor (2005) did measure parenting quality in terms of variables like parental talkativeness and conversational responsiveness, and found that it was a less important predictor of 2-year-olds ToM performance than was positive sibling relationship quality. Cole and Mitchell (2000) postulated that the effects of a sibling on children s ToM performance might be mediated through a more basic effect of siblings on children s EF capacities for planning, flexibility and inhibitory control. We tested this mediation hypothesis. However, despite significant associations among key variables, there was no statistically reliable evidence of this mediation effect. Within the limits of our sample and the particular EF and ToM tests we used, the influence of child siblings on ToM did not appear to be fully channelled through executive cognition. Furthermore, our data hinted at subtle differences in the manner in which the sibling interaction opportunity variable was linked with individual differences in EF performance, in contrast to ToM performance, for these preschoolers. While both of the child sibling variables we identified, namely (a) the presence of at least one child-aged sibling, and (b) the total number of child siblings in the family, were linked with ToM performance, links with EF scores showed a slightly different pattern. The number of child siblings correlated with EF scores both before and after verbal ability was partialled out, whereas the correlation with the presence of a child-aged sibling only achieved statistical significance once verbal ability was statistically controlled. If having more than just one child sibling turns out to be more important for EF performance than for ToM (a possibility that requires further empirical investigation), this could have implications for theories of how child siblings influence preschoolers social and cognitive development. While speculative, one possibility is that the total number of child playmates relates linearly to advanced EF skills because group interaction with several siblings at once is of special value in cultivating EF. For example, teamwork may be critical, as when a group of siblings engages in competitive or strategic games. For ToM, on the other hand, dyadic conversation and pretend play with just one sibling may also supply developmentally useful input, assisting the child s understanding both of other people s roles and perspectives and of the subjectivity of inner mental states of believing or pretending. Further research that directly addresses these questions using observations of family interaction would be valuable, and additional empirical evidence is clearly needed before such possibilities can be anything other than highly speculative. Also in need of further research is the question of how EF and ToM development interconnect with one another over the course of development. Hughes and Ensor s (2005) findings provide important indications of correlations that pre-date the preschool period, as well as of the importance of high quality interaction among siblings for both EF and ToM capacities emerging during toddlerhood, and these patterns are largely consistent with present findings. However, the accumulated empirical evidence to date still does not permit a definitive choice among existing theoretical models for how EF relates to ToM. These, as Perner and Lang (1999) noted, fall into three broad groupings. Some theories postulate that EF abilities are prerequisites for the development of ToM. Others view ToM as a prerequisite that is needed before EF can

18 750 Anna McAlister and Candida C. Peterson develop, whereas theories of a third kind postulate that both types of skill may develop simultaneously, either through coordinated maturation of brain structures and/or through shared dependency on similar socializing experiences and supporting skills (e.g. language). After reviewing the empirical evidence then available, Perner and Lang (1999) concluded that each of these theoretical possibilities was tenable, and the present findings continue to align themselves with the same open-ended causal picture. Yet, our results do make an important addition to the picture by demonstrating that the social influence of an available child playmate in the family is beneficial to EF performance, as well as well as to ToM, during the preschool period. It is to be hoped that future research, ideally incorporating both longitudinal and naturalistic-observational assessment, can further refine the understanding both of the specific manner in which interaction among child siblings may influence ToM and EF performance, and vice versa. Future research, ideally incorporating a longitudinal design, is also needed to clarify the developmental paths of change and interdependency among these core cognitive and social-cognitive abilities. References Astington, J. A. (2003). Sometimes necessary, never sufficient: False belief understanding and social competence. In B. Repacholi & V. Slaughter (Eds.), Individual differences in theory of mind ( pp ). New York: Psychology Press. Baron, R., & Kenny, D. (1986). The moderator-mediator distinction in social psychological research. Journal of Personality and Social Psychology, 51, Baron-Cohen, S., Leslie, A. M., & Frith, U. (1985). Does the autistic child have a theory of mind. Cognition, 21, Brown, J., Donelan-McCall, N., & Dunn, J. (1996). Why talk about mental states? The significance of children s conversations with friends, siblings, and mothers. Child Development, 67, Brown, J., & Dunn, J. (1992). Talk with your mother or your sibling: Developmental changes in early family conversations about feelings. Child Development, 63, Carlson, S. M., & Moses, L. J. (2001). Individual differences in inhibitory control and children s theory of mind. Child Development, 72, Carlson, S. M., Moses, L. J., & Hix, H. R. (1998). The role of inhibitory processes in young children s difficulties with deception and false belief. Child Development, 69, Cole, K., & Mitchell, P. (2000). Siblings in the development of executive control and a theory of mind. British Journal of Developmental Psychology, 18, Cutting, A. L., & Dunn, J. (1999). Theory of mind, emotion understanding, language, and family background: Individual differences and interrelations. Child Development, 70, Dunn, J. (1994). Changing minds and changing relationships. In C. Lewis & P. Mitchell (Eds.), Origins of an understanding of mind ( pp ). Hove, UK: Erlbaum. Dunn, J. (1995). Children as psychologists: The later correlates of individual differences in understanding emotions and other minds. Cognition and Emotion, 9, Dunn, L. M., & Dunn, L. M. (1981). Peabody Picture Vocabulary Test-Revised. Circle Pines, MN: American Guidance Service. Flavell, J. H. (1986). The development of children s knowledge about the appearance-reality distinction. American Psychologist, 41, Flavell, J. H. (1999). Cognitive development: Children s knowledge about the mind. Annual Review of Psychology, 50, Glass, D., Neulinger, J., & Brim, O. G. (1974). Birth order verbal intelligence and educational aspirations. Child Development, 45,

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