Laterality: Asymmetries of Body, Brain and Cognition
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Need for cognition moderates paranormal beliefs and magical ideation in inconsistent-handers Eric C. Prichard & Stephen D. Christman To cite this article: Eric C. Prichard & Stephen D. Christman (2016): Need for cognition moderates paranormal beliefs and magical ideation in inconsistent-handers, Laterality: Asymmetries of Body, Brain and Cognition, DOI: 10.1080/1357650X.2015.1125914 To link to this article: http://dx.doi.org/10.1080/1357650X.2015.1125914
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Date: 19 February 2016, At: 23:31
LATERALITY: ASYMMETRIES OF BODY, BRAIN AND COGNITION, 2016 http://dx.doi.org/10.1080/1357650X.2015.1125914
Need for cognition moderates paranormal beliefs and magical ideation in inconsistent-handers Downloaded by [University of California Santa Barbara] at 23:31 19 February 2016
Eric C. Prichard and Stephen D. Christman Department of Psychology, University of Toledo, Toledo, OH, USA ABSTRACT
A growing literature suggests that degree of handedness predicts gullibility and magical ideation. Inconsistent-handers (people who use their non-dominant hand for at least one common manual activity) report more magical ideation and are more gullible. The current study tested whether this effect is moderated by need for cognition. One hundred eighteen university students completed questionnaires assessing handedness, self-reported paranormal beliefs, and self-reported need for cognition. Handedness (Inconsistent vs. Consistent Right) and Need for Cognition (High vs. Low) were treated as categorical predictors. Both paranormal beliefs and magical ideation served as dependent variable’s in separate analyses. Neither set of tests yielded main effects for handedness or need for cognition. However, there were a significant handedness by need for cognition interactions. Post-hoc comparisons revealed that low, but not high, need for cognition inconsistenthanders reported relatively elevated levels of paranormal belief and magical ideation. A secondary set of tests treating the predictor variables as continuous instead of categorical obtained the same overall pattern. ARTICLE HISTORY Received 14 August 2015; Accepted 25 November 2015 KEYWORDS Handedness; magical ideation; paranormal beliefs; need for cognition
An expanding corpus of research supports the notion that degree of handedness in an important individual difference variable because it may serve as a proxy for access to the right hemisphere. A recent review by Prichard, Propper, and Christman (2013) discusses findings demonstrating that inconsistenthandedness (as opposed to consistent/strong handedness) predicts better performance on episodic memory tasks and more frequent belief updating and cognitive flexibility, independent of direction of handedness. Both of these findings are consistent with theories which postulate right hemisphere roles in episodic retrieval (Habib, Nyberg, & Tulving, 2003) and belief updating (Ramachandran, 1995). However, little research to date has examined CONTACT Eric C. Prichard © 2016 Taylor & Francis
[email protected]
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psychological moderators of these handedness effects. The purpose of the study which follows is to investigate whether another psychological variable, need for cognition (Cacioppo & Petty, 1982; Cacioppo, Petty, & Kao, 1984), moderates handedness as a predictor of variables associated with cognitive flexibility and belief updating.
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Handedness and belief updating Ramachandran’s (1995) work with patients who developed anosognosia after suffering from right temporal lesions led him to propose that the left hemisphere maintains one’s present schemas, while the right hemisphere acts as a “devil’s advocate,” which detects anomalies and forces the left hemisphere to update its beliefs in response to new information. Hence, patients with right temporal damage do not update their beliefs in response to new information and occasionally deny their conditions. Operating under the framework that consistency of handedness is associated with greater access to the right hemisphere, based on (i) larger corpus callosum size in inconsistent-handers (Luders et al., 2010) compared to consistent right handers and (ii) greater resting right hemisphere activation among inconsistent right handers as compared to consistent right handers (Propper, Pierce, Geisler, Christman, & Bellorado, 2012), researchers have investigated whether inconsistently handed people have a greater tendency to update beliefs in response to new information. Early correlational evidence was supportive of this possibility. Barnett and Corballis (2002) showed that people with inconsistent hand preference reported higher scores on an inventory measuring magical ideation, a finding replicated by Nicholls, Orr, and Lindell (2005).1 Magical ideation, which is a predictor of future schizotypy, is the tendency to believe in causal relationships which are considered impossible by conventional standards (Eckblad & Chapman, 1983), and schizotypy is also associated with inconsistent-handedness (e.g., Asai & Tanno, 2009; Poreh, Levin, Teves, & States, 1997; Tran, Stieger, & Voracek, 2015). More direct evidence comes from several laboratory studies designed to directly address the question. Christman, Henning, Geers, Propper, and Niebauer (2008) conducted a set of studies in which they looked at handedness as a predictor of persuadability and gullibility. In their first study, they demonstrated that inconsistent-handers showed more sensitivity to arguments in favour of a student comprehensive exam (a hypothetical exam one would have to pass in order to obtain a bachelor’s degree). In the second study, inconsistent-handers showed more susceptibility to the Barnum effect, 1
However, it is worth noting that some researchers question whether the relationship between handedness and magical ideation has a neuroanatomical basis (e.g. Badzakova-Trajkov, Häberling, & Corballis, 2011; Grimshaw, Yelle, Schoger, & Bright, 2008).
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which is the tendency to believe that a generic evaluation which could describe anyone is an accurate profile of one’s own specific personality. Belief in the Barnum effect can be argued to be a measure of gullibility. The gullibility finding has been extended to demonstrate that inconsistenthanders show more cognitive dissonance than consistent-handers after receiving false personality profiles accusing them of sexism (Jasper, Prothero, & Christman, 2009). This was taken to be evidence of greater belief updating, because inconsistent-handers would have to be more likely to believe bogus feedback that challenged their self-schemas in order to experience more cognitive dissonance in response to accusations of sexism. In addition there have been findings showing that inconsistent-handers are also more prone to the placebo effect, which is thought to be driven by one’s belief that a treatment will work (Christman, Geers, Kosbab, & Weiland, 2006). Further evidence for increased belief updating and cognitive flexibility in inconsistent-handers has been found in the domains of musical preferences (Christman, 2013), body image perception (Niebauer, Aselage, & Schutte, 2002), ambiguous figure perception (Christman, Sontam, & Jasper, 2009), and judgment and decision-making (Jasper, Barry, & Christman, 2008; Jasper, Kunzler, Prichard, & Christman, 2014). The purpose of the present study is to take a step towards identifying possible moderators of the tendency for inconsistent-handers to update their beliefs more readily. The issue here centres on an apparent contradiction: given that inconsistent-handers are more gullible, it makes sense that they would be more prone to magical ideation. On the other hand, given that inconsistent-handers are more persuadable and cognitively flexible, why are they then not open to scientific evidence against magical and paranormal phenomena? Beliefs in magical and paranormal phenomena were measured separately. Although there is some overlap between the constructs of paranormal beliefs and magical ideation, we believe that paranormal beliefs as measured by the scale used is a better indicator of acceptance of pseudoscientific ideas while magical ideation is a better predictor of acceptance of cognitions predictive of schizotypy. This distinction is important because schizotypy as measured by magical ideation is a predictor of serious organically based psychotic illnesses such as schizophrenia (Eckblad & Chapman, 1983). Holding paranormal beliefs does not require abnormal neurophysiological morphology or abnormal neurotransmitter activity. However, someone with more magical ideation might also have more paranormal beliefs as a result of their predisposition for psychosis. In this way, schizotypy and paranormal beliefs may have a high correlation, yet different aetiologies in different subpopulations, making them distinct constructs. Preliminary analyses showed that the two variables were correlated, r = .652, p < .001.
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Thus, using a paranormal beliefs scale and a scale measuring magical ideation as proxies for belief in ideas supported by ambiguous evidence, the authors set out to test whether need for cognition moderates the effects on handedness on belief updating. Need for cognition is the tendency for people to enjoy cognitively effortful activity. Perhaps the handedness difference in magical ideation is driven by inconsistent-handers who are low in need for cognition, which might prevent them being exposed to and/or deeply processing counter-evidence. High need for cognition inconsistenthanders, on the other hand, might be initially prone to magical and paranormal beliefs but would, thanks to their high need for cognition, eventually encounter counter-evidence and update their beliefs accordingly. Need for cognition is the tendency for people to enjoy cognitively demanding activity. If need for cognition were shown to attenuate or exacerbate the tendency for inconsistent-handers to accept beliefs for which there is scant evidence, it would provide an important qualifier to existing research. It would also raise the question of which inconsistent-handers update their beliefs and when. It was hypothesized that, if divided into high vs. low need for cognition, low need for cognition inconsistent-handers would report higher paranormal beliefs and magical ideation than the other three groups. High need for cognition inconsistent-handers were not expected to differ from consistenthanders.
Methods Participants and procedure One hundred eighteen university students (96 females) participated in the study for course credit. The participants ranged in age from 15 to 38 (M = 18.87, SD = 2.48). Up to two participants completed the study at a time. Upon entering the data collection room, participants were greeted by an experimenter. The experimenter obtained consent from the participants. Upon obtaining consent, the experimenter instructed the participants to complete a questionnaire packet. Once the questionnaires were completed, participants were granted course credit.
Materials The 18-Item Need for Cognition Scale The 18-Item Need for Cognition Scale (NFC) (Cacioppo et al., 1984) is a shortened version of the NFC (Cacioppo & Petty, 1982). The scale is used as a measure of enjoyment for activities which require a high level of cognitive engagement. The scale consists of seven-point Likert-type items (e.g., “I find
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satisfaction in deliberating hard and for long hours.”) with anchors at −3 (Strongly Disagree) and 3 (Strongly Agree).
The Magical Ideation Scale The Magical Ideation Scale (MIS) (Eckblad & Chapman, 1983) is a 30 item instrument which consists of true false items which were designed to measure schizotypy. Magical ideation predicts risk for schizophrenia and schizotypal disorders, both of which are often characterized by magical thinking. The items consist of forced choice true/false statements regarding magical beliefs (e.g., “Some people can make me aware of them just by thinking about me”). A high score indicate the endorsement of a high number of magical beliefs.
The Revised Paranormal Beliefs Scale The Revised Paranormal Beliefs Scale (RPBS) (Tobacyk, 2004) is a 26-item scale designed to measure paranormal beliefs which are common in Western culture. While it can be divided into subscales (looking at belief in religion, psi, witchcraft, superstition, spiritualism, extraordinary lifeforms, and precognition) the scale is used here as a global assessment of paranormal beliefs. The scale consists of seven-point Likert-type item (e.g., “Black magic really exists.”) with anchors at 1 (Strongly Disagree) and 7 (Strongly Agree).
The Edinburgh Handedness Inventory The present study used a slightly modified version of the Edinburgh Handedness Inventory (EHI) (Oldfield, 1971). The inventory presented 10 common motor tasks. Specifically, the version used assessed hand preference for writing, drawing, using a spoon, opening jars, using a toothbrush, throwing, combing one’s hair, using scissors, using a knife, and striking a match. As demonstrated by Edlin et al. (2015), there are multiple versions of the EHI in use and researchers often fail to specify which items their versions use. This is an important point, since several versions of the EHI differ slightly from one another and from the original Oldfield (1971) version. It is not yet fully known how slight differences in scales referred to as “The EHI” may affect the generalizability of handedness findings. For this reason, we have listed the items from the version used in this study. Furthermore, the response format differs slightly from the original Oldfield versions. Response options fell along a five-point Likert-type scale ranging from Always Left (−10) to Always Right (+10). The maximum score was +100. The minimum raw score was −100. When absolute values are used, scores range from 0 to 100.
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Results One participant was dropped from the analysis for failing to complete all of the questionnaires. The remaining 117 participants were divided into inconsistent and consistent-handed groups by taking the absolute values of the EHI and performing a median split. The median absolute EHI score was 80. Individuals with scores of 80 were included in the inconsistent-handed group in order to allow for the most balanced cell sizes possible. No left handed participants had an absolute EHI score above 80. For that reason, consistent-handers are referred to in this report as consistent right handers. While we believe degree of handedness is an important variable, it is still possible that consistent left handers could vary from consistent right handers and inconsistent-handers under some circumstances (e.g., Hardie & Wright, 2014). Participants were also divided into a high need for cognition and low need for cognition groups by taking a median split of the 18-Item NFC. The median score was 8. In order to balance cell sizes to the greatest extent possible, we placed participants with scores of 8 and below were placed in the “Low Need for Cognition” group. Preliminary analyses of the NFC showed that consistent right handers (M = 3.72, SD = 23.58) and inconsistenthanders (M = 9.75, SD = 23.63) did not statistically differ in need for cognition, t(115) = 1.37, p = .172. Data were analysed using a 2 (Hand: Inconsistent vs. Consistent) × 2 (Need for Cognition: Low vs. High) analysis of covariance. Because some evidence suggests that certain components of schizotypal thinking may tend to become manifest later in adolescence (e.g., Fonseca-Pedrero, Lemos-Giraldez, Muniz, Garcia-Cueto, & Campillo-Alvarez, 2008) and because young college aged adults may still be formulating attitudes about the paranormal and critical thinking, age was included as a covariate. This was deemed especially important given that our participants tended to be older adolescents and young adults. The categorization process yielded four cells: Low Need for Cognition Inconsistent-Handers (N = 29), High Need for Cognition InconsistentHanders (N = 29), Low Need for Cognition Consistent Right Handers (N = 34) and High Need for Cognition Consistent Right Handers (N = 25). The first series of analyses used the RPBS as the criterion variable and handedness and need for cognition as the categorical predictors. There were no main effects of handedness or need for cognition. However, there was a significant handedness by need for cognition interaction, F(1, 113) = 5.99, p = .016, m2partial = .051 (see Figure 1). The interaction was broken down by conducting least squared difference tests on the cells. The decision was made to use more liberal tests after conducting a post-hoc power analysis using GPower. Given the estimates effect sizes reported below, all of our paired comparisons were underpowered if the power criterion of .8 is used. In the time since Cohen (1962) first raised
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Figure 1. Paranormal beliefs (RPBS): Handedness by need for cognition interaction.
the issue of Type II errors resulting from underpowered studies, research has shown that even real effects of medium effect size should go undetected fairly frequently given an alpha level of .05 and the use of recommended power levels (e.g., Francis, 2013). The interaction was driven by low need for cognition inconsistent-handers (M = 92.45, SE = 4.5) scoring higher on paranormal beliefs than high need for cognition inconsistent-handers (M = 77.62, SE = 4.2), p = .011, d = .59 and low need for cognition inconsistent-handers scoring higher than low need for cognition consistent-handers (M = 75.76, SE = 4.5), p = .018, d = .68. The second series of analyses used the MIS as the criterion variable, while handedness and need for cognition served as the categorical predictor variables. There were no main effects of handedness or of need for cognition, which is inconsistent with previous literature showing that inconsistent-handedness is associated with more magical ideation (Barnett & Corballis, 2002). However, there was a significant handedness by need for cognition interaction, F(1, 113) = 4.18, p = .043, m2partial = .036 (see Figure 2). The interaction was broken down by conducting least squared difference tests on the cells. The rationale for a more liberal test was similar to the rationale used when choosing pairwise tests to that elucidated for the tests reported above. The interaction was driven by the tendency for low need for cognition inconsistent-handers (M = 10.48, SE = .88) to answer “true” to more items than high need for cognition inconsistent-handers (M = 9.35, SE = .81), p = .051, d = .22.
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Figure 2. Magical ideation: Handedness by need for cognition interaction.
Secondary analyses: treating all variables as continuous Because some researchers (e.g., MacCallum, Zhang, Preacher, & Rucker, 2002) object to the dichotomization of continuous variables and suggest that such a practice may distort findings, we did a secondary analysis in which we treated degree of handedness and need for cognition as continuous variables. Absolute EHI scores, Need for Cognition scores, age, and the absolute EHI by need for cognition interaction term were included as predictor variables. We also include tables showing the means, standard deviations, and a simple correlation matrix of the five variables under study (see Tables 1 and 2). When the RPBS served as the criterion variable, the overall model was significant, F(4, 111) = 2.43, p = .051, R 2 = .081. This effect was driven by need for cognition and the handedness by need for cognition interaction acting as significant predictors of paranormal beliefs (see Table 3). In order to interpret in the interaction, we constructed a graph of the simple slopes (see Figure 3). The graph shows that at one standard deviation below the mean absolute EHI score, individuals with a need for cognition scores one standard deviation below the mean would be predicted to have a paranormal beliefs score (Predicted RPBS = 85.86) higher than the predicted paranormal beliefs score of individuals with a need for cognition score one standard deviation above the need for cognition mean (Predicted RPBS = 37.46). There is a crossover at 1 SD above the mean for handedness, with a score of 1 SD above the handedness mean predicting a higher paranormal beliefs score for participants
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who are at 1 SD above the NFC mean (Predicted RPBS = 77.78) than for participants who are at 1 SD below the mean for NFC (Predicted RPBS = 67.81). Similarly, when magical ideation served as the criterion variable, the overall model was significant, F(4, 111) = 4.31, p = .003, R2 = .135. All four predictor variables were significant predictors of magical ideation (see Table 4). Once again, a graph of simple slopes was constructed (see Figure 4).The graph shows that at one standard deviation below the mean absolute EHI score, individuals with NFC scores 1 SD below the mean would be predicted to have magical ideation score (Predicted Magical Ideation = 4.76) higher than the predicted magical ideation score of participants with NFC scores 1 SD above the need for cognition mean (Predicted Magical Ideation = 2.04). There is a crossover at 1 SD above the mean for handedness, with a score of hand 1 SD above the handedness mean predicting a higher magical ideation scores for participants who are at 1 SD above the NFC mean (Predicted Magical Ideation = 4.37) than for participants who are at 1 SD below the mean for NFC (Predicted Magical Ideation = 0.59). The pattern of data is essentially the same whether the predictors are treated as dichotomized categorical predictors in an analysis of variance or continuous predictors in a regression model. Inconsistent-handers tend to have lower scores on measures of paranormal beliefs and magical ideation than do consistent right handers when these individuals score low on measures of magical ideation. The pattern seems to reverse when need for cognition is high.
Discussion Although consistent right handers and inconsistent-handers did not statistically differ on the RPBS or the MIS, ANCOVAs controlling for age yielded significant handedness by need for cognition interactions when the RPBS and the MIS served as criterion variables. Low need for cognition inconsistenthanders scored higher on paranormal beliefs than high need for cognition inconsistent-handers and low need for cognition consistent right handers. Interestingly, high need for cognition consistent right handers scored nominally higher on paranormal beliefs and magical ideation than low need for cognition consistent-handers. However, these differences did not reach the level of statistical significance, so it would be unwise to over-interpret this finding. Table 1. Descriptives. Variable
Mean
Standard deviation
Absolute EHI scores 18-Item Need for Cog Scale Paranormal (RPBS) Magical ideation (MIS)
77.09 6.97 82.23 9.49
22.87 23.70 25.04 4.77
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Table 2. Pearson correlation coefficients. Variables
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RPBS MIS Absolute EHI Need for Cog **p < .01.
RPBS
MIS
Absolute EHI
Need for Cog
1.00 0.652** −0.063 −0.031
1.00 −0.090 −0.095
1.00 −0.135
1.00
The finding that high need for cognition seems to attenuate paranormal beliefs and magical ideation for inconsistent-handers is valuable because it suggests that enjoyment of engaging in thought reduces the probability that a group which may be susceptible to scantily supported beliefs will accept said beliefs. However, it also raises several questions which need to be addressed in future studies. First, we can offer some speculation as to the temporal trajectory of magical ideation and paranormal beliefs. Presumably, people do not start out believing in the paranormal. Perhaps arising from the increased cognitive flexibility and belief updating observed in inconsistent-handers, most inconsistent-handers initially embrace such beliefs upon first encountering them. Then, the high need for cognition inconsistent-handers, but not the low need for cognition ones, eventually seek out or come across disconfirming information that allows a second round of belief updating in which the paranormal beliefs are now rejected. The low need for cognition inconsistenthanders do not seek out such information and hence do not update their beliefs. Magical ideation could have a slightly different trajectory. We argue that magical ideation is a symptom of schizotypy, which may have a biological basis. As such, people higher in schizotypy may develop thoughts such as “people can read my mind” or “I experience precognition in dreams.” It might be that inconsistent-handers initially accept these cognitions because they have more cognitive flexibility or that inconsistent-handedness and magical ideation are spuriously related and both caused by a third variable. In either case, we would suspect that high need for cognition inconsistenthanders seek out additional information which leads to the rejection of these cognition. Hence, “People are reading my mind” might become “People are reading my mind … but wait, I have read there is no evidence for extrasensory perception (ESP). I am just imagining it.” Low need for Table 3. Regression analysis one (model predicting RPBS scores). Intercept Absolute EHI Need for Cog Age AbsEHI × NFC
B
SE of B
β
Significance
88.62 −0.173 −0.1089 0.456 0.013
18.95 0.107 0.375 0.934 0.004
−0.159 −1.035 0.045 1.015
.000 .107 .003 .626 .003
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Figure 3. Handedness by need for cognition simple slopes: Paranormal beliefs (RPBS). Table 4. Regression analysis two (model predicting magical ideation). Intercept Absolute EHI Need for Cog Age AbsEHI × NFC
B
SE of B
β
Significance
6.015 −0.041 −0.220 0.359 0.003
3.52 0.020 0.066 0.174 0.001
−0.196 −1.092 0.187 1.199
.090 .041 .001 .041 .000
Figure 4. Handedness by need for cognition simple slopes: Magical ideation.
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cognition inconsistent-handers may never gain access to information which would allow them to reject magical cognitions. In contrast, our study yielded weak evidence that consistent-handers systematically differ in magical ideation or paranormal beliefs as a function of need for cognition. It might be that once beliefs are set, whether they be for against the paranormal, very little may change for consistent-handers even when they have access to additional information. Similarly, perhaps when consistent-handers develop schizotypal ideations, it will be more difficult to reject those cognitions even when given evidence that such cognitions are internally generated and do not reflect a violation of standard rules of causality in our universe. Given the descriptive crossover interaction, it may even be worth considering and testing the hypothesis that high need for condition consistent-handers do seek out information, but primarily confirming information. Second, does need for cognition result in an overall decline in belief updating for inconsistent-handers, or does it depend of the belief in question? For example, inconsistent-handers are more likely to believe in evolution (Niebauer, Christman, Reid, & Garvey, 2004), a theory for which there is much support. As such, one might imagine that high need for cognition inconsistent-handers are the ones who would be more likely to support the theory of evolution. It is also unknown how need for cognition will moderate other belief updating findings. Are high need for cognition inconsistent-handers more resilient to the Barnum effect and the placebo effect? What about perceptual illusions? Furthermore, our sample of consistent-handers was exclusively right handed. Would the pattern of results differ for a group of consistent left handers? Future studies should seek samples of consistent left handers and determine whether there are circumstances where consistency and direction of handedness interact. Third, future research might attempt to disentangle the separate components of magical ideation vs. paranormal beliefs. These two measures are typically highly correlated, and the current paper treats them basically as two alternative measures of non-normative beliefs. However, it is worth noting that need for cognition had a stronger effect on inconsistenthanders’ paranormal beliefs than it did on their magical ideation. This may reflect the fact that paranormal beliefs in things such as UFOs and ESP are more amenable to empirical disconfirmation than are magical ideations such as “I have occasionally had the silly feeling that a TV or radio broadcaster knew I was listening to him” and “The hand motions that strangers make seems to influence me at times.” In other words, paranormal beliefs may reflect increased gullibility to exogenous information (e.g., Christman et al., 2008), while magical ideation may reflect a more endogenous form of disordered thought. In this sense, the association between handedness and schizotypy may be driven more by magical ideation than paranormal beliefs.
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Finally, researchers should investigate more potential moderators of the handedness effects observed thus far in the literature. As more is learned about consistency of handedness as a predictor, it will be important to study how consistency of handedness interacts with other variables and traits that predict behaviour. The study of categorical predictor variables carries the risk of oversimplifying individual differences. The field of laterality has done much to come back from failed pseudo-psychological conceptions such as “left brainers vs. right brainers.” The study of moderators associated with handedness will allow the field to develop a nuanced view of the consistency of handedness which will allow researchers to know when the main effects reported in previous findings hold and when they do not, which in turn will make consistency of handedness an even richer and more informative variable for researchers.
Acknowledgement We would like to acknowledge Kritika Singh for assisting with data collection and data entry.
Disclosure statement No potential conflict of interest was reported by the authors.
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Interhemispheric interaction and belief updating. Laterality, 13, 403–426. doi:10. 1080/13576500802079646 Christman, S. D., Sontam, V., & Jasper, J. D. (2009). Individual differences in ambiguous figure perception: Degree of handedness and interhemispheric interaction. Perception, 38, 1183–1198. doi:10.1068/p6131 Cohen, J. (1962). The statistical power of abnormal-social psychological research. Journal of Abnormal and Social Psychology, 65, 145–153. doi:10.1037/h0045186 Eckblad, M., & Chapman, L. J. (1983). Magical ideation as an indicator of schizotypy. Journal of Consulting and Clinical Psychology, 51, 215–225. doi:10.1037/0022-006X. 51.2.215 Edlin, J. M., Leppanen, M. L., Fain, R. J., Hackländer, R. P., Hanaver-Torrez, S. D., & Lyle, K. B. (2015). On the use (and misuse?) of the Edinburgh Handedness Inventory. Brain and Cognition, 94, 44–51. doi:10.1016/j.bandc.2015.01.003 Fonseca-Pedrero, E., Lemos-Giraldez, S., Muniz, J., Garcia-Cueto, E., & Campillo-Alvarez, A. (2008). Schizotpy in adolescence: The role of gender and age. The Journal of Nervous and Mental Disease, 196, 161–165. doi:10.1097/NMD.0b013e318162aa79 Francis, G. (2013). Replication, statistical consistency, and publication bias. Journal of Mathematical Psychology, 57, 153–169. doi:10.1016/j.jmp.2013.02.003 Grimshaw, G. M., Yelle, S. K., Schoger, J., & Bright, K. S. (2008). Magical ideation is related to questionnaire but not behavioural measures of handedness. Laterality, 13, 22–33. doi:10.1080/13576500701508539 Habib, R., Nyberg, L., & Tulving, E. (2003). Hemispheric asymmetries of memory: The HERA model revisited. Trends in Cognitive Sciences, 7, 241–245. doi:10.1016/S13646613(03)00110-4 Hardie, S. M., & Wright, L. (2014) Differences between left- and right-handers in approach/avoidance motivation: Influence of consistency of handedness measures. Frontiers in Psychology, 5, 134. doi:10.3389/fpsyg.2014.00134 Jasper, J. D., Barry, K., & Christman, S. D. (2008). Individual differences in counterfactual production. Personality and Individual Differences, 45, 488–492. doi:10.1016/j.paid. 2008.05.026 Jasper, J. D., Kunzler, J., Prichard, E. C., & Christman, S. D. (2014). Individual differences in information order effects: The importance of right-hemisphere access in belief updating. Acta Psychologica, 148, 115–122. doi:10.1016/j.actpsy.2014.01.004 Jasper, J. D., Prothero, M., & Christman, S. D. (2009). I’m not sexist!!! Cognitive dissonance and the differing cries of mixed-and strong-handers. Personality and Individual Differences, 47, 268–272. doi:10.1016/j.paid.2009.03.010 Luders, E., Cherbuin, N., Thompson, P. M., Gutman, B., Anstey, K. J., Sachdev, P., & Toga, A. W. (2010). When more is less: Associations between corpus callosum size and handedness lateralization. NeuroImage, 52, 43–49. doi:10.1016/j.neuroimage.2010. 04.016 MacCallum, R. C., Zhang, S., Preacher, K. J., & Rucker, D. D. (2002). On the practice of dichotomization of quantitative variables. Psychological Methods, 7, 19–40. doi:10. 1037/1082-989X.7.1.19 Nicholls, M. E. R., Orr, C. A., & Lindell, A. K. (2005). Magical ideation and its relation to lateral preference. Laterality, 10, 503–515. doi:10.1080/13576500442000265 Niebauer, C., Aselage, J., & Schutte, C. (2002). Interhemispheric interaction and consciousness: Degree of handedness predicts the intensity of a sensory illusion. Laterality, 7, 85–96. doi:10.1080/13576500143000159
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Niebauer, C., Christman, S., Reid, S., & Garvey, K. (2004). Interhemispheric interaction and beliefs on our origin: Degree of handedness predicts beliefs in creationism versus evolution. Laterality, 9, 433–447. doi:10.1080/13576500342000266 Oldfield, R. C. (1971). The assessment and analysis of handedness: The Edinburgh inventory. Neuropsychologia, 9, 97–113. doi:10.1016/0028-3932(71)90067-4 Poreh, A. M., Levin, J., Teves, H., & States, J. (1997). Mixed handedness and schizotypal personality in a non-clinical sample – the role of task demand. Personality and Individual Differences, 23, 501–507. doi:10.1016/S0191-8869(97)00057-3 Prichard, E., Propper, R. E., & Christman, S. D. (2013). Degree of handedness, but not direction, is a systematic predictor of cognitive performance. Frontiers in Psychology, 4, 3–6. doi:10.3389/fpsyg.2013.00009 Propper, R. E., Pierce, J., Geisler, M. W., Christman, S. D., & Bellorado, N. (2012). Asymmetry in resting alpha activity: Effects of handedness. Open Journal of Medical Psychology, 1, 86–90. doi:10.4236/ojmp.2012.14014 Ramachandran, V. S. (1995). Anosognosia in parietal lobe syndrome. Consciousness and Cognition, 4, 22–51. doi:10.1006/ccog.1995.1002 Tobacyk, J. J. (2004). A revised paranormal beliefs scale. International Journal of Transpersonal Studies, 23, 94–98. Tran, U. S., Stieger, S., & Voracek, M. (2015). Mixed-footedness is a more relevant predictor of schizotypy than mixed-handedness. Psychiatry Research, 225, 446–451. doi:10.1016/j.psychres.2014.11.069
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Need for cognition moderates paranormal beliefs and magical ideation in inconsistent-handers Eric C. Prichard & Stephen D. Christman To cite this article: Eric C. Prichard & Stephen D. Christman (2016): Need for cognition moderates paranormal beliefs and magical ideation in inconsistent-handers, Laterality: Asymmetries of Body, Brain and Cognition, DOI: 10.1080/1357650X.2015.1125914 To link to this article: http://dx.doi.org/10.1080/1357650X.2015.1125914
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Need for cognition moderates paranormal beliefs and magical ideation in inconsistent-handers Downloaded by [University of California Santa Barbara] at 23:31 19 February 2016
Eric C. Prichard and Stephen D. Christman Department of Psychology, University of Toledo, Toledo, OH, USA ABSTRACT
A growing literature suggests that degree of handedness predicts gullibility and magical ideation. Inconsistent-handers (people who use their non-dominant hand for at least one common manual activity) report more magical ideation and are more gullible. The current study tested whether this effect is moderated by need for cognition. One hundred eighteen university students completed questionnaires assessing handedness, self-reported paranormal beliefs, and self-reported need for cognition. Handedness (Inconsistent vs. Consistent Right) and Need for Cognition (High vs. Low) were treated as categorical predictors. Both paranormal beliefs and magical ideation served as dependent variable’s in separate analyses. Neither set of tests yielded main effects for handedness or need for cognition. However, there were a significant handedness by need for cognition interactions. Post-hoc comparisons revealed that low, but not high, need for cognition inconsistenthanders reported relatively elevated levels of paranormal belief and magical ideation. A secondary set of tests treating the predictor variables as continuous instead of categorical obtained the same overall pattern. ARTICLE HISTORY Received 14 August 2015; Accepted 25 November 2015 KEYWORDS Handedness; magical ideation; paranormal beliefs; need for cognition
An expanding corpus of research supports the notion that degree of handedness in an important individual difference variable because it may serve as a proxy for access to the right hemisphere. A recent review by Prichard, Propper, and Christman (2013) discusses findings demonstrating that inconsistenthandedness (as opposed to consistent/strong handedness) predicts better performance on episodic memory tasks and more frequent belief updating and cognitive flexibility, independent of direction of handedness. Both of these findings are consistent with theories which postulate right hemisphere roles in episodic retrieval (Habib, Nyberg, & Tulving, 2003) and belief updating (Ramachandran, 1995). However, little research to date has examined CONTACT Eric C. Prichard © 2016 Taylor & Francis
[email protected]
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psychological moderators of these handedness effects. The purpose of the study which follows is to investigate whether another psychological variable, need for cognition (Cacioppo & Petty, 1982; Cacioppo, Petty, & Kao, 1984), moderates handedness as a predictor of variables associated with cognitive flexibility and belief updating.
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Handedness and belief updating Ramachandran’s (1995) work with patients who developed anosognosia after suffering from right temporal lesions led him to propose that the left hemisphere maintains one’s present schemas, while the right hemisphere acts as a “devil’s advocate,” which detects anomalies and forces the left hemisphere to update its beliefs in response to new information. Hence, patients with right temporal damage do not update their beliefs in response to new information and occasionally deny their conditions. Operating under the framework that consistency of handedness is associated with greater access to the right hemisphere, based on (i) larger corpus callosum size in inconsistent-handers (Luders et al., 2010) compared to consistent right handers and (ii) greater resting right hemisphere activation among inconsistent right handers as compared to consistent right handers (Propper, Pierce, Geisler, Christman, & Bellorado, 2012), researchers have investigated whether inconsistently handed people have a greater tendency to update beliefs in response to new information. Early correlational evidence was supportive of this possibility. Barnett and Corballis (2002) showed that people with inconsistent hand preference reported higher scores on an inventory measuring magical ideation, a finding replicated by Nicholls, Orr, and Lindell (2005).1 Magical ideation, which is a predictor of future schizotypy, is the tendency to believe in causal relationships which are considered impossible by conventional standards (Eckblad & Chapman, 1983), and schizotypy is also associated with inconsistent-handedness (e.g., Asai & Tanno, 2009; Poreh, Levin, Teves, & States, 1997; Tran, Stieger, & Voracek, 2015). More direct evidence comes from several laboratory studies designed to directly address the question. Christman, Henning, Geers, Propper, and Niebauer (2008) conducted a set of studies in which they looked at handedness as a predictor of persuadability and gullibility. In their first study, they demonstrated that inconsistent-handers showed more sensitivity to arguments in favour of a student comprehensive exam (a hypothetical exam one would have to pass in order to obtain a bachelor’s degree). In the second study, inconsistent-handers showed more susceptibility to the Barnum effect, 1
However, it is worth noting that some researchers question whether the relationship between handedness and magical ideation has a neuroanatomical basis (e.g. Badzakova-Trajkov, Häberling, & Corballis, 2011; Grimshaw, Yelle, Schoger, & Bright, 2008).
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which is the tendency to believe that a generic evaluation which could describe anyone is an accurate profile of one’s own specific personality. Belief in the Barnum effect can be argued to be a measure of gullibility. The gullibility finding has been extended to demonstrate that inconsistenthanders show more cognitive dissonance than consistent-handers after receiving false personality profiles accusing them of sexism (Jasper, Prothero, & Christman, 2009). This was taken to be evidence of greater belief updating, because inconsistent-handers would have to be more likely to believe bogus feedback that challenged their self-schemas in order to experience more cognitive dissonance in response to accusations of sexism. In addition there have been findings showing that inconsistent-handers are also more prone to the placebo effect, which is thought to be driven by one’s belief that a treatment will work (Christman, Geers, Kosbab, & Weiland, 2006). Further evidence for increased belief updating and cognitive flexibility in inconsistent-handers has been found in the domains of musical preferences (Christman, 2013), body image perception (Niebauer, Aselage, & Schutte, 2002), ambiguous figure perception (Christman, Sontam, & Jasper, 2009), and judgment and decision-making (Jasper, Barry, & Christman, 2008; Jasper, Kunzler, Prichard, & Christman, 2014). The purpose of the present study is to take a step towards identifying possible moderators of the tendency for inconsistent-handers to update their beliefs more readily. The issue here centres on an apparent contradiction: given that inconsistent-handers are more gullible, it makes sense that they would be more prone to magical ideation. On the other hand, given that inconsistent-handers are more persuadable and cognitively flexible, why are they then not open to scientific evidence against magical and paranormal phenomena? Beliefs in magical and paranormal phenomena were measured separately. Although there is some overlap between the constructs of paranormal beliefs and magical ideation, we believe that paranormal beliefs as measured by the scale used is a better indicator of acceptance of pseudoscientific ideas while magical ideation is a better predictor of acceptance of cognitions predictive of schizotypy. This distinction is important because schizotypy as measured by magical ideation is a predictor of serious organically based psychotic illnesses such as schizophrenia (Eckblad & Chapman, 1983). Holding paranormal beliefs does not require abnormal neurophysiological morphology or abnormal neurotransmitter activity. However, someone with more magical ideation might also have more paranormal beliefs as a result of their predisposition for psychosis. In this way, schizotypy and paranormal beliefs may have a high correlation, yet different aetiologies in different subpopulations, making them distinct constructs. Preliminary analyses showed that the two variables were correlated, r = .652, p < .001.
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Thus, using a paranormal beliefs scale and a scale measuring magical ideation as proxies for belief in ideas supported by ambiguous evidence, the authors set out to test whether need for cognition moderates the effects on handedness on belief updating. Need for cognition is the tendency for people to enjoy cognitively effortful activity. Perhaps the handedness difference in magical ideation is driven by inconsistent-handers who are low in need for cognition, which might prevent them being exposed to and/or deeply processing counter-evidence. High need for cognition inconsistenthanders, on the other hand, might be initially prone to magical and paranormal beliefs but would, thanks to their high need for cognition, eventually encounter counter-evidence and update their beliefs accordingly. Need for cognition is the tendency for people to enjoy cognitively demanding activity. If need for cognition were shown to attenuate or exacerbate the tendency for inconsistent-handers to accept beliefs for which there is scant evidence, it would provide an important qualifier to existing research. It would also raise the question of which inconsistent-handers update their beliefs and when. It was hypothesized that, if divided into high vs. low need for cognition, low need for cognition inconsistent-handers would report higher paranormal beliefs and magical ideation than the other three groups. High need for cognition inconsistent-handers were not expected to differ from consistenthanders.
Methods Participants and procedure One hundred eighteen university students (96 females) participated in the study for course credit. The participants ranged in age from 15 to 38 (M = 18.87, SD = 2.48). Up to two participants completed the study at a time. Upon entering the data collection room, participants were greeted by an experimenter. The experimenter obtained consent from the participants. Upon obtaining consent, the experimenter instructed the participants to complete a questionnaire packet. Once the questionnaires were completed, participants were granted course credit.
Materials The 18-Item Need for Cognition Scale The 18-Item Need for Cognition Scale (NFC) (Cacioppo et al., 1984) is a shortened version of the NFC (Cacioppo & Petty, 1982). The scale is used as a measure of enjoyment for activities which require a high level of cognitive engagement. The scale consists of seven-point Likert-type items (e.g., “I find
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satisfaction in deliberating hard and for long hours.”) with anchors at −3 (Strongly Disagree) and 3 (Strongly Agree).
The Magical Ideation Scale The Magical Ideation Scale (MIS) (Eckblad & Chapman, 1983) is a 30 item instrument which consists of true false items which were designed to measure schizotypy. Magical ideation predicts risk for schizophrenia and schizotypal disorders, both of which are often characterized by magical thinking. The items consist of forced choice true/false statements regarding magical beliefs (e.g., “Some people can make me aware of them just by thinking about me”). A high score indicate the endorsement of a high number of magical beliefs.
The Revised Paranormal Beliefs Scale The Revised Paranormal Beliefs Scale (RPBS) (Tobacyk, 2004) is a 26-item scale designed to measure paranormal beliefs which are common in Western culture. While it can be divided into subscales (looking at belief in religion, psi, witchcraft, superstition, spiritualism, extraordinary lifeforms, and precognition) the scale is used here as a global assessment of paranormal beliefs. The scale consists of seven-point Likert-type item (e.g., “Black magic really exists.”) with anchors at 1 (Strongly Disagree) and 7 (Strongly Agree).
The Edinburgh Handedness Inventory The present study used a slightly modified version of the Edinburgh Handedness Inventory (EHI) (Oldfield, 1971). The inventory presented 10 common motor tasks. Specifically, the version used assessed hand preference for writing, drawing, using a spoon, opening jars, using a toothbrush, throwing, combing one’s hair, using scissors, using a knife, and striking a match. As demonstrated by Edlin et al. (2015), there are multiple versions of the EHI in use and researchers often fail to specify which items their versions use. This is an important point, since several versions of the EHI differ slightly from one another and from the original Oldfield (1971) version. It is not yet fully known how slight differences in scales referred to as “The EHI” may affect the generalizability of handedness findings. For this reason, we have listed the items from the version used in this study. Furthermore, the response format differs slightly from the original Oldfield versions. Response options fell along a five-point Likert-type scale ranging from Always Left (−10) to Always Right (+10). The maximum score was +100. The minimum raw score was −100. When absolute values are used, scores range from 0 to 100.
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Results One participant was dropped from the analysis for failing to complete all of the questionnaires. The remaining 117 participants were divided into inconsistent and consistent-handed groups by taking the absolute values of the EHI and performing a median split. The median absolute EHI score was 80. Individuals with scores of 80 were included in the inconsistent-handed group in order to allow for the most balanced cell sizes possible. No left handed participants had an absolute EHI score above 80. For that reason, consistent-handers are referred to in this report as consistent right handers. While we believe degree of handedness is an important variable, it is still possible that consistent left handers could vary from consistent right handers and inconsistent-handers under some circumstances (e.g., Hardie & Wright, 2014). Participants were also divided into a high need for cognition and low need for cognition groups by taking a median split of the 18-Item NFC. The median score was 8. In order to balance cell sizes to the greatest extent possible, we placed participants with scores of 8 and below were placed in the “Low Need for Cognition” group. Preliminary analyses of the NFC showed that consistent right handers (M = 3.72, SD = 23.58) and inconsistenthanders (M = 9.75, SD = 23.63) did not statistically differ in need for cognition, t(115) = 1.37, p = .172. Data were analysed using a 2 (Hand: Inconsistent vs. Consistent) × 2 (Need for Cognition: Low vs. High) analysis of covariance. Because some evidence suggests that certain components of schizotypal thinking may tend to become manifest later in adolescence (e.g., Fonseca-Pedrero, Lemos-Giraldez, Muniz, Garcia-Cueto, & Campillo-Alvarez, 2008) and because young college aged adults may still be formulating attitudes about the paranormal and critical thinking, age was included as a covariate. This was deemed especially important given that our participants tended to be older adolescents and young adults. The categorization process yielded four cells: Low Need for Cognition Inconsistent-Handers (N = 29), High Need for Cognition InconsistentHanders (N = 29), Low Need for Cognition Consistent Right Handers (N = 34) and High Need for Cognition Consistent Right Handers (N = 25). The first series of analyses used the RPBS as the criterion variable and handedness and need for cognition as the categorical predictors. There were no main effects of handedness or need for cognition. However, there was a significant handedness by need for cognition interaction, F(1, 113) = 5.99, p = .016, m2partial = .051 (see Figure 1). The interaction was broken down by conducting least squared difference tests on the cells. The decision was made to use more liberal tests after conducting a post-hoc power analysis using GPower. Given the estimates effect sizes reported below, all of our paired comparisons were underpowered if the power criterion of .8 is used. In the time since Cohen (1962) first raised
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Figure 1. Paranormal beliefs (RPBS): Handedness by need for cognition interaction.
the issue of Type II errors resulting from underpowered studies, research has shown that even real effects of medium effect size should go undetected fairly frequently given an alpha level of .05 and the use of recommended power levels (e.g., Francis, 2013). The interaction was driven by low need for cognition inconsistent-handers (M = 92.45, SE = 4.5) scoring higher on paranormal beliefs than high need for cognition inconsistent-handers (M = 77.62, SE = 4.2), p = .011, d = .59 and low need for cognition inconsistent-handers scoring higher than low need for cognition consistent-handers (M = 75.76, SE = 4.5), p = .018, d = .68. The second series of analyses used the MIS as the criterion variable, while handedness and need for cognition served as the categorical predictor variables. There were no main effects of handedness or of need for cognition, which is inconsistent with previous literature showing that inconsistent-handedness is associated with more magical ideation (Barnett & Corballis, 2002). However, there was a significant handedness by need for cognition interaction, F(1, 113) = 4.18, p = .043, m2partial = .036 (see Figure 2). The interaction was broken down by conducting least squared difference tests on the cells. The rationale for a more liberal test was similar to the rationale used when choosing pairwise tests to that elucidated for the tests reported above. The interaction was driven by the tendency for low need for cognition inconsistent-handers (M = 10.48, SE = .88) to answer “true” to more items than high need for cognition inconsistent-handers (M = 9.35, SE = .81), p = .051, d = .22.
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Figure 2. Magical ideation: Handedness by need for cognition interaction.
Secondary analyses: treating all variables as continuous Because some researchers (e.g., MacCallum, Zhang, Preacher, & Rucker, 2002) object to the dichotomization of continuous variables and suggest that such a practice may distort findings, we did a secondary analysis in which we treated degree of handedness and need for cognition as continuous variables. Absolute EHI scores, Need for Cognition scores, age, and the absolute EHI by need for cognition interaction term were included as predictor variables. We also include tables showing the means, standard deviations, and a simple correlation matrix of the five variables under study (see Tables 1 and 2). When the RPBS served as the criterion variable, the overall model was significant, F(4, 111) = 2.43, p = .051, R 2 = .081. This effect was driven by need for cognition and the handedness by need for cognition interaction acting as significant predictors of paranormal beliefs (see Table 3). In order to interpret in the interaction, we constructed a graph of the simple slopes (see Figure 3). The graph shows that at one standard deviation below the mean absolute EHI score, individuals with a need for cognition scores one standard deviation below the mean would be predicted to have a paranormal beliefs score (Predicted RPBS = 85.86) higher than the predicted paranormal beliefs score of individuals with a need for cognition score one standard deviation above the need for cognition mean (Predicted RPBS = 37.46). There is a crossover at 1 SD above the mean for handedness, with a score of 1 SD above the handedness mean predicting a higher paranormal beliefs score for participants
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who are at 1 SD above the NFC mean (Predicted RPBS = 77.78) than for participants who are at 1 SD below the mean for NFC (Predicted RPBS = 67.81). Similarly, when magical ideation served as the criterion variable, the overall model was significant, F(4, 111) = 4.31, p = .003, R2 = .135. All four predictor variables were significant predictors of magical ideation (see Table 4). Once again, a graph of simple slopes was constructed (see Figure 4).The graph shows that at one standard deviation below the mean absolute EHI score, individuals with NFC scores 1 SD below the mean would be predicted to have magical ideation score (Predicted Magical Ideation = 4.76) higher than the predicted magical ideation score of participants with NFC scores 1 SD above the need for cognition mean (Predicted Magical Ideation = 2.04). There is a crossover at 1 SD above the mean for handedness, with a score of hand 1 SD above the handedness mean predicting a higher magical ideation scores for participants who are at 1 SD above the NFC mean (Predicted Magical Ideation = 4.37) than for participants who are at 1 SD below the mean for NFC (Predicted Magical Ideation = 0.59). The pattern of data is essentially the same whether the predictors are treated as dichotomized categorical predictors in an analysis of variance or continuous predictors in a regression model. Inconsistent-handers tend to have lower scores on measures of paranormal beliefs and magical ideation than do consistent right handers when these individuals score low on measures of magical ideation. The pattern seems to reverse when need for cognition is high.
Discussion Although consistent right handers and inconsistent-handers did not statistically differ on the RPBS or the MIS, ANCOVAs controlling for age yielded significant handedness by need for cognition interactions when the RPBS and the MIS served as criterion variables. Low need for cognition inconsistenthanders scored higher on paranormal beliefs than high need for cognition inconsistent-handers and low need for cognition consistent right handers. Interestingly, high need for cognition consistent right handers scored nominally higher on paranormal beliefs and magical ideation than low need for cognition consistent-handers. However, these differences did not reach the level of statistical significance, so it would be unwise to over-interpret this finding. Table 1. Descriptives. Variable
Mean
Standard deviation
Absolute EHI scores 18-Item Need for Cog Scale Paranormal (RPBS) Magical ideation (MIS)
77.09 6.97 82.23 9.49
22.87 23.70 25.04 4.77
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Table 2. Pearson correlation coefficients. Variables
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RPBS MIS Absolute EHI Need for Cog **p < .01.
RPBS
MIS
Absolute EHI
Need for Cog
1.00 0.652** −0.063 −0.031
1.00 −0.090 −0.095
1.00 −0.135
1.00
The finding that high need for cognition seems to attenuate paranormal beliefs and magical ideation for inconsistent-handers is valuable because it suggests that enjoyment of engaging in thought reduces the probability that a group which may be susceptible to scantily supported beliefs will accept said beliefs. However, it also raises several questions which need to be addressed in future studies. First, we can offer some speculation as to the temporal trajectory of magical ideation and paranormal beliefs. Presumably, people do not start out believing in the paranormal. Perhaps arising from the increased cognitive flexibility and belief updating observed in inconsistent-handers, most inconsistent-handers initially embrace such beliefs upon first encountering them. Then, the high need for cognition inconsistent-handers, but not the low need for cognition ones, eventually seek out or come across disconfirming information that allows a second round of belief updating in which the paranormal beliefs are now rejected. The low need for cognition inconsistenthanders do not seek out such information and hence do not update their beliefs. Magical ideation could have a slightly different trajectory. We argue that magical ideation is a symptom of schizotypy, which may have a biological basis. As such, people higher in schizotypy may develop thoughts such as “people can read my mind” or “I experience precognition in dreams.” It might be that inconsistent-handers initially accept these cognitions because they have more cognitive flexibility or that inconsistent-handedness and magical ideation are spuriously related and both caused by a third variable. In either case, we would suspect that high need for cognition inconsistenthanders seek out additional information which leads to the rejection of these cognition. Hence, “People are reading my mind” might become “People are reading my mind … but wait, I have read there is no evidence for extrasensory perception (ESP). I am just imagining it.” Low need for Table 3. Regression analysis one (model predicting RPBS scores). Intercept Absolute EHI Need for Cog Age AbsEHI × NFC
B
SE of B
β
Significance
88.62 −0.173 −0.1089 0.456 0.013
18.95 0.107 0.375 0.934 0.004
−0.159 −1.035 0.045 1.015
.000 .107 .003 .626 .003
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Figure 3. Handedness by need for cognition simple slopes: Paranormal beliefs (RPBS). Table 4. Regression analysis two (model predicting magical ideation). Intercept Absolute EHI Need for Cog Age AbsEHI × NFC
B
SE of B
β
Significance
6.015 −0.041 −0.220 0.359 0.003
3.52 0.020 0.066 0.174 0.001
−0.196 −1.092 0.187 1.199
.090 .041 .001 .041 .000
Figure 4. Handedness by need for cognition simple slopes: Magical ideation.
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cognition inconsistent-handers may never gain access to information which would allow them to reject magical cognitions. In contrast, our study yielded weak evidence that consistent-handers systematically differ in magical ideation or paranormal beliefs as a function of need for cognition. It might be that once beliefs are set, whether they be for against the paranormal, very little may change for consistent-handers even when they have access to additional information. Similarly, perhaps when consistent-handers develop schizotypal ideations, it will be more difficult to reject those cognitions even when given evidence that such cognitions are internally generated and do not reflect a violation of standard rules of causality in our universe. Given the descriptive crossover interaction, it may even be worth considering and testing the hypothesis that high need for condition consistent-handers do seek out information, but primarily confirming information. Second, does need for cognition result in an overall decline in belief updating for inconsistent-handers, or does it depend of the belief in question? For example, inconsistent-handers are more likely to believe in evolution (Niebauer, Christman, Reid, & Garvey, 2004), a theory for which there is much support. As such, one might imagine that high need for cognition inconsistent-handers are the ones who would be more likely to support the theory of evolution. It is also unknown how need for cognition will moderate other belief updating findings. Are high need for cognition inconsistent-handers more resilient to the Barnum effect and the placebo effect? What about perceptual illusions? Furthermore, our sample of consistent-handers was exclusively right handed. Would the pattern of results differ for a group of consistent left handers? Future studies should seek samples of consistent left handers and determine whether there are circumstances where consistency and direction of handedness interact. Third, future research might attempt to disentangle the separate components of magical ideation vs. paranormal beliefs. These two measures are typically highly correlated, and the current paper treats them basically as two alternative measures of non-normative beliefs. However, it is worth noting that need for cognition had a stronger effect on inconsistenthanders’ paranormal beliefs than it did on their magical ideation. This may reflect the fact that paranormal beliefs in things such as UFOs and ESP are more amenable to empirical disconfirmation than are magical ideations such as “I have occasionally had the silly feeling that a TV or radio broadcaster knew I was listening to him” and “The hand motions that strangers make seems to influence me at times.” In other words, paranormal beliefs may reflect increased gullibility to exogenous information (e.g., Christman et al., 2008), while magical ideation may reflect a more endogenous form of disordered thought. In this sense, the association between handedness and schizotypy may be driven more by magical ideation than paranormal beliefs.
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Finally, researchers should investigate more potential moderators of the handedness effects observed thus far in the literature. As more is learned about consistency of handedness as a predictor, it will be important to study how consistency of handedness interacts with other variables and traits that predict behaviour. The study of categorical predictor variables carries the risk of oversimplifying individual differences. The field of laterality has done much to come back from failed pseudo-psychological conceptions such as “left brainers vs. right brainers.” The study of moderators associated with handedness will allow the field to develop a nuanced view of the consistency of handedness which will allow researchers to know when the main effects reported in previous findings hold and when they do not, which in turn will make consistency of handedness an even richer and more informative variable for researchers.
Acknowledgement We would like to acknowledge Kritika Singh for assisting with data collection and data entry.
Disclosure statement No potential conflict of interest was reported by the authors.
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