.

.

However, neuroimaging experiments can sample brain activ-ity systematically and noninvasively in healthy volunteers (Pol-drack and Farah, 2015) and, with appropriate methodologies,they can also provide valuable information about the functionalspecificity of brain regions (Moran and Zaki, 2013; Poldrack andFarah, 2015).

.

Together, stimulation studies support the idea that some pos-terior cortical regions may contribute directly to specific contentsof experience, but the evidence for prefrontal regions is scarceand indirect.

.

.

.

Electrical stimulation during neurosurgery is an important source ofevidence for a direct contribution of different brain areas to con-sciousness (Penfield, 1959; Desmurget et al., 2013), as indicatedby its superior value in predicting postoperative deficits com-pared with fMRI or diffusion tensor imaging (Borchers et al.,2011).

Although frontal injuries can slightlyincrease the threshold for perceiving some brief (16 ms) andmasked visual stimuli, patients still experience them (Del Cul etal., 2009), suggesting that these frontal regions may modulate theNCC (i.e., act as background conditions) rather than contribut-ing directly to consciousness (Kozuch, 2014).

By contrast, there is little evidence for loss of specific con-scious contents after frontal damage (Penfield and Jasper, 1954).

With regards to content-specific NCC, there is abundant neu-rological evidence that lesions in the posterior cortex can cause aloss of specific contents of experience (Farah, 2004).

.

.

For content-specific NCC, experimentscan be carefully designed to systematically investigate possibledissociations between the experience of particular conscious con-tents and the engagement of various cognitive processes, such asattention, decision-making, and reporting (Aru et al., 2012; Kochand Tsuchiya, 2012; Tsuchiya et al., 2015; Tsuchiya and Koch,2016).

Several complementary methods can be used to distill the trueNCC. For the full NCC, within-state paradigms can be used toavoid confounds due to changes in behavioral state and taskperformance as well as to dissociate unconsciousness from unre-sponsiveness

.

The full NCC can be definedas the union of all content-specific NCC (Koch et al., 2016a).

scious percept (Crick and Koch, 1990). Content-specific NCCare the neural mechanisms specifying particular phenomenalcontents within consciousness, such as colors, faces, places, orthoughts.

The neural correlates of consciousness (NCC) are defined as theminimal neural mechanisms jointly sufficient for any one con-

.

The role of the frontal cortex in consciousness remains a matter of debate. In this Perspective, we will critically review the clinical andneuroimaging evidence for the involvement of the front versus the back of the cortex in specifying conscious contents and discusspromising research avenues.

Consistent with these reports,we found no differences in the frequency of these percepts between the first and second half of the experiment (Z = 0.085, p = .932,r = 0.013).

We additionally examined whether the frequency of these experiences varied throughout the experiment to clarify additionalelements of this phenomenon.

Participants demonstrated a significantly higher frequency of visual percepts onbeep trials (Z = 1.992, p = .046, r = 0.307; Fig. 1A), consistent with our suggestion that startling tones presented in the context ofmild visual deprivation can lead to auditory-visual percepts.

We first examined whether it is possible to evoke auditory-visual synesthesia in non-synesthetes undergoing short-term sensorydeprivation and parameterized the features that maximize the strength of these experiences (Klüver, 1966).

.

Conversely, the disinhibited-feedback model proposes that everyone possesses the potential to experience synesthesia, mani-festing only when the balance of activity across the senses has been altered (Grossenbacher & Lovelace, 2001). According to this view,the anatomical structure of the brains of synesthetes and non-synesthetes is generally similar, but there are differences in howeffectively one sense can evoke activity in a second modality, potentially due to weaker inhibition of feedback projections thatconnect the sensory systems (Grossenbacher & Lovelace, 2001). For example, instead of abnormal connections enabling sounds toevoke conscious visual experiences in synesthetes (as in the cross-activation model), this model argues that these connections arepresent in all individuals and that synesthetes have less inhibition restricting how strongly sounds modulate visual activity.

If all individuals possess the ability for one sensory system to modulate activity in another, why do only a minority of individualsexperience synesthesia? The cross-activation model reconciles this discrepancy by arguing that synesthesia arises from anatomicalpathways that are either weak or absent in non-synesthetes, providing a direct link through which one sensory modality can stimulateanother modality (Hubbard, Arman, Ramachandran, & Boynton, 2005; Hubbard, Brang, & Ramachandran, 2011; Ramachandran andHubbard, 2001a, 2001b, 2003).

.

.

Competing models differ in the time required for synesthetic experiences to emerge. Thecross-activation model suggests synesthesia arises over months or years from the development ofabnormal neural connections. Here we demonstrate that after ∼5 min of visual deprivation,sounds can evoke synesthesia-like percepts (vivid colors and Klüver form-constants) in ∼50% ofnon-synesthetes. These results challenge aspects of the cross-activation model and suggest thatsynesthesia exists as a latent feature in all individuals, manifesting when the balance of activityacross the senses has been altered.

.

.

Using this approach,a recent study found that spectral power in the delta band in posterior cortex was higher during reported uncon-sciousness than during reported consciousness 23 . Furthermore, using a within-state design in NREM sleep, ithas been found that TMS triggers a larger negative EEG peak amplitude during reported unconsciousness thanduring reported consciousness, indicating that differences in consciousness within the same physiological stateare related to local alterations in the cortical bistability of posterior brain regions 21.

Previous studies have mostly compared wakefulness with sleep or anesthesia to evaluate features associatedwith the level of consciousness in healthy individuals 14,18,19 . However, such studies are confounded by otherchanges that occur across global state shifts, such as changes in the cardiovascular, respiratory, and neuromus-cular systems 20 . To address this issue, recent studies have measured the presence or absence of consciousnesswithin the same physiologically categorized state using a within-state paradigm21–23.

.

Together these resultssuggest that alterations in spectral and spatial characteristics of network properties in posterior brainareas, in particular decreased local (segregated) connectivity at low frequencies, is a potential indicatorof consciousness during sleep.

The neuronal connectivity patterns that differentiate consciousness from unconsciousness remainunclear. Previous studies have demonstrated that effective connectivity, as assessed by transcranialmagnetic stimulation combined with electroencephalography (TMS–EEG), breaks down duringthe loss of consciousness.

.

.

.

.

.

We aimed to investigate the risk of substance use-relatedproblems in ASD. We also tested if any association betweenASD and substance use-related problems could be related tocomorbidity with ADHD or intellectual disability (ID). Toelucidate if shared familial factors underlie both ASD andsubstance use-related problems, we examined the pattern ofsubstance use-related problems also among unaffected rela-tives of individuals with ASD.

.

In other words, register-based stud-ies are likely to underestimate the absolute prevalence ofsubstance use disorder since users not in contact with thetreatment system are not taken into account (EMCDDA2004).

Second, we could stratify our analysis on comorbidADHD and ID only when those disorders were diagnosed,but it impossible to rule out that patients with seemingly“pure” ASD have undiagnosed ID, ADHD, or subthresh-old ADHD-symptoms involving substance use-relatedproblems

.

Strengths include the large scale population-based design,prospectively collected data from nationwide registries,stratification by comorbid disorders, statistical control forsocio-demographic confounders, and analysis of familialaggregation data from relatives. Nevertheless, some limita-tions deserve comments.

.

We can only speculate that the same familial factors maybe causal in substance use-related problems among ASDprobands may lead also to higher risks of substance-relateddeath among their non-ASD relatives.

As a result, it turned out that patients with ASD only andASD with ADHD are actually on comparable risks of sub-stance use-related problems (OR 1.6 vs. 1.9) and previouslydescribed extremely high risk in patents with ASD andADHD seems to be due diagnostic biases.

.

.

.

.

.

1. to examine stated reasons for initiation of andrelapse to substance use,2. to examine reasons and strategies used for quit-ting, and3. to explore the perceived association betweensubstance use and mental illness among a largesample of persons with co-occurring SMI andSUD.

The aims of this articleare

Fishbein’s(1980) theory of reasoned action postulates that be-havior is based on attitudes that, in turn, are based onpersonal beliefs. Beliefs rest in large part on what islearnt and experienced; in particular, beliefs that arebased on personal experience have been found tohave a stronger influence in the formation of attitudesthan information gained in other ways and to betterpredict later behavior (Fazio & Zanna, 1981).

The supersensitivity model, wherebybiological vulnerability due to psychiatric disorderresults in sensitivity to small amounts of alcoholand drugs, leading to substance misuse, has alsoreceived some support (e.g., Lieberman, Kane, &Alvir, 1987).

.

.

.

The etiology of substance use among persons with severe mental illness remains unclear. Thisstudy investigates stated reasons for substance use among persons in recovery from co-occurringdisorders of serious mental illness and substance abuse and dependence. The desire to fit in withpeers played a key role in the initiation of substance use; boredom, loneliness, temptations to use,and stress were cited most as relapse triggers.

.

Increased risk of substance use-related problems seemsto contradict global negative attitudes towards psychoac-tive substances observed among ASD patients (Ramos etal. 2013). Individuals with ASD may find them helpful toreduce tension and enhance social skills more often thannon-ASD controls do (Cludius et al. 2013).

Despite limited and ambiguous empirical data,substance use-related problems have been assumed to berare among patients with autism spectrum disorders (ASD).

.

We found that ASD was associated with increased risk fora range of substance use-related problems, and the familydata suggested that this was due to shared liability betweenASD and substance use-related problems between relatives.

We conclude that ASD is arisk factor for substance use-related problems. The elevatedrisks among relatives of probands with ASD suggest sharedfamilial (genetic and/or shared environmental) liability.

.

To address these limitations, in Experiment 1, we requiredparticipants to rate their confidence after each individual response(rather than after each pair of responses).

In Experiment 4, we manipulated participants’ confidence priorto administration of the MRT. All participants first completed aline judgment task that was intentionally difficult, so that par-ticipants would be unable to gauge their performance.

Upon completion of the line judgment task, participants wererandomly informed that their performance on the line judgmenttask was either above average (‘‘high confidence’’condition) orbelow average (‘‘low confidence’’ condition).

Ifconfidence mediates mental rotation performance, then partic-ipants in the high confidence condition should outperform theircounterparts in the low confidence condition.

Cooke-Simpson and Voyer (2007) provided tentative evi-dence that confidence predicted MRT performance, but thatstudy had several critical limitations.

correlation was comparable to that observed in prior studies(r = ?.69; Cooke-Simpson & Voyer, 2007).

As illustrated in Fig. 2, confidence predicted accuracy acrossboth sexes, r(67) = ?.56, p\.001, and the strength of this

Thisresult replicates the general pattern observed in Experiment 1(Fig. 3), and indicates that both males’ and females’ confi-dence was indeed calibrated to their performance.

Experiment3 therefore supported the hypothesis that mental rotation ismediated by confidence.

Experiment 3 provided a further test of whether the sex differencein performance is better explained by confidence or by omissions.

Results are summarized in Table 1. The sex difference in accu-racy was replicated in the omission condition but not in the com-mission condition.

Thus, the sex difference in mental rotationwas attributable to confidence rather than omissions.

mediates mental rotation. If confidence was unrelated to mentalrotation, then the sex difference should be equivalent acrossgroups (i.e., no interaction should occur).

Critically, an inter-action in either direction would suggest that confidence

.

.

In Experiment 2, we sought to attenuate the sex difference inmental rotation performance by rendering confidence irrelevantto the task.

Fig. 4

In conclusion, Experiment 1 corroborated the finding thatconfidence predicted mental rotation performance both acrossand within sexes (Cooke-Simpson & Voyer, 2007). Experiment1 further demonstrated, for the first time, that confidence pre-dicted mental rotation performance within individuals: Partic-ipants were more accurate on trials for which they were moreconfident. These results thus provide the most precise evidenceto date of the relation between confidence and mental rotation.Finally, Experiment 1 also provided the first evidence of thedirection of this relationship: Mediation analyses revealed thatconfidence mediated the sex difference in mental rotation per-formance whereas mental rotation performance did not mediatethe sex difference in confidence.

If confidence mediates mental rotation performance,then confidence ought to predict accuracy on the MRT acrosssexes, within each sex, and possibly even within individuals.

In Experiment 1, we tested whether confidence predicted men-tal rotation performance between sexes, within each sex, andwithin individuals.

Thus, mental rotation performance did notmediate the sex difference in confidence, but rather confidencestrongly mediated the sex difference in mental rotation per-formance.

So, together, Figs. 2 and 3 reveal thatmales and females were similarly successful at calibrating theirconfidence to their accuracy (Fig. 3), though males tendedtoward the upper part of the distribution on both confidence andaccuracy (Fig. 2).

As expected, an independent samples t-test revealed that males(M = 5.61, SD = 1.02) were more confident than females (M =4.62, SD = 1.41), d = .74, t(65) = 3.26, p\.01.

Specifically, if confidence medi-ates mental rotation, then (1) confidence should predict mentalrotation scores not only between sexes, but also within sexes, (2)rendering confidence irrelevant to the task should attenuate thesex difference, and (3) manipulating participants’ confidenceshould affect their mental rotation performance.

.

Because confidence appears tobe an important component of the MRT, it stands as a plausiblemediator of performance.

The most common measure of mental rotation performanceis the Mental Rotations Test (MRT; Vandenberg & Kuse, 1978),which is based on the 3-dimensional block figures introduced byShepard and Metzler (1971) and updated by Peters et al. (1995).

.

Preliminary evidence suggests that confidence might indeedunderlie the sex difference in mental rotation.

.

We therefore tested whether confidence mediatedmental rotation performance.

basic cognitive skills, such as attention, memory, and judgment(e.g., Schmader et al., 2008), which ultimately would affectperformance.

The beliefthat one (or one’s social group) is skilled or poor at a given taskmay well affect one’s confidence when approaching that taskand this effect on confidence may have cascading effects on

Much of the research on gender role and sex stereotype effectsassumes confidence as a potential cognitive mechanism bywhich those social factors exert their effect.

So, in summary, beliefs about and aware-ness of sex stereotypes are both related to the sex difference inmental rotation performance. But how exactly might sex ste-reotypes affect performance?

Mental rotation performance may also be affected by mereawareness of, rather than belief in, the stereotype that men aresuperior to women on spatial tasks.

.

Performance on spatialtasks thus is clearly related to gender role beliefs and traits.

Gender role beliefs and traits may partially explain the sex dif-ference in mental rotation performance.

Here, we examined whether one suchsociocognitive factor, namely participants’ confidence, contrib-uted to this sex difference in mental rotation performance.Although this presumed relation between confidence and men-tal rotation performance has received little empirical attention,related research on gender roles, sex stereotypes, and stereotypethreat provides a rich source of supportive evidence.

.

Given the complexity of the taskand the magnitude of the sex difference, it likely has multi-ple causes or mediators.

Of all cognitive sex differences, the mental rotation of abstractfigures in 3-dimensional space is the most robust (Halpern, 2000;Hines, 2004; Linn & Petersen, 1985; Maccoby & Jacklin, 1974).

Thus, confidence medi-ates the sex difference in mental rotation performance and hencethe sex difference appears to be a difference of performancerather than ability. Results are discussed in relation to otherpotential mediators and mechanisms, such as gender roles,sex stereotypes, spatial experience, rotation strategies, work-ing memory, and spatial attention.

On tasks that require the mental rotation of 3-dimensional figures, males typically exhibit higher accuracythan females. Using the most common measure of mental rota-tion (i.e., the Mental Rotations Test), we investigated whetherindividual variability in confidence mediates this sex differ-ence in mental rotation performance.

.

Thus, the third approach to the issue of Sex differencesand the time factor in the MRT examined the sex differ-ence in a RT paradigm where only two mental rotationfigures were used.

.

For this reason, Study 2, which manipulatestime directly, compares sex differences under the stan-dard condition with sex differences which are observedwhen time is increased, but within limits.

Here, we administered the MRT under identical con-ditions, but allowing two durations.

.

If time pressure is a significant factor in per-formance, we expect to see two indicators in the data.First, we expect to see that females attempt significantlyfewer problems than males and, second, we expect to seethat the magnitude of the sex difference increases as theproblem position increases.

In the present study, three different approaches to theproblem of time constraints are taken, each examining adifferent aspect of how time might affect the sex differ-ences on the MRT.

Thus, our understanding of the role of time con-straints in MRT sex differences remains inconclusive.

Several studies find that when enough time is pro-vided for the V&K MRT, sex differences on the MRTdisappear (Goldstein, Haldane, & Mitchell, 1990;Voyer, 1997), leading to the conclusion that sex differ-ences arise because the sexes differ in the amount of timetaken to perform the mental rotation.

The idea that the activational effects of hormones mightlead to relatively faster mental rotations touches uponthe time factor in MRT sex differences.

That males and females differ in performance on Van-denberg and Kuse (1978) mental rotation task is wellknown (Voyer, Voyer, & Bryden, 1995). The causes areless well understood.

.

We conclude that performancefactors may play a role in sex difference on mental rotation tasks, but do not account for all of the differences.

In accounting for the well-established sex differences on mental rotation tasks that involve cube stimuli of the Shepard and Met-zler (Shepard & Metzler, 1971) kind, performance factors are frequently invoked.

.

Gender differences in favor of men in spatial abilities are one of the most commonly replicated finding inpsychology research (Hedges & Nowell, 1995; Linn & Petersen, 1985; Voyer, Voyer, & Bryden, 1995). Among thetasks that produce such differences, the Mental Rotations Test (MRT), developed by Vandenberg and Kuse (1978),produces the largest effects (Linn & Petersen, 1985; Peters, 2005).

Eighty undergraduate students (40 males, 40 females) completed the MRT while ratingtheir confidence in the accuracy of their answers for each item. As expected, gender differences in favor of men were obtained.Results also indicated a positive correlation between confidence ratings and scores on the MRT, as well as negative correlationsbetween confidence ratings and MRT outcomes presumed to reflect propensity to guess. More elaborate analyses using a measureof accuracy of predictions (the Brier score) indicated that men have a more accurate perception of their performance on the MRTthan women do.

Further support for a constant anduniform biological explanation comes from studies that findsimilar effect sizes for gender across cultures (Silverman,Phillips, & Silverman, 1996).

The distinct hormonal environments of men and womenmay play a role.

.

.

When brain imaging studies are done on subjects per-forming mental rotation tasks, there is some evidence ofactivation in motor areas of the brain.

.

Biological factors have also been discussed as potentialcauses for this difference. Biological theories stress theimportance of genetics, hormonal influence, brain organi-zation, and maturational factors.

There is,however, no consensus as to the role of task complexity ongender differences in mental rotation.

Task difficulty is one potential performance factor thathas been explored (Bryden, George, & Inch, 1990; Collins& Kimura, 1997).

.

.

A number of explanations have been advanced for theexistence of the gender difference in mental rotation.

Voyer, Voyer, & Bryden’s (1995) meta-analysis of sexdifferences in spatial abilities, found that the average differ-ence (using Cohen’s d = (M1 − M2)/σ) between men andwomen on the (MRT; Vandenberg & Kuse, 1978), was 0.94(this represents a very large effect), indicating that men per-form nearly one standard deviation above the average per-formance of women.

The ability to mentally rotate an object has been foundto produce one of the largest sex differences in the cogni-tive literature (Linn & Petersen, 1985).

The visuospatial ability referred to as mental rotation has been shown to produce one of the largest and most consistent sex differences,in favor of males, in the cognitive literature.

Sex differences were also seen in the patterns of correlations between rotation tasks and other neuropsycho-logical measures. Current results suggest men may rely more on left hemisphere processing than women when engaged in rotational tasks.© 2003 Elsevier Ltd. All rights reserved.

.

.

.

Pre-vious results with similar stimuli and mixed groups of participants(Han, Humphreys, & Chen, 1999; Kimchi, 1994) showed the globaladvantage that is typically observed with hierarchical stimuli (e.g.Kimchi, 1992; Navon, 1977), as well as faster classification by clo-sure than by line orientation. In addition, the global advantage wasmore pronounced when local classification was based on line ori-entation than on closure (Han et al., 1999; Kimchi, 1994).

Thus, in this study, we examined gender differences in global–local processing with a visuospatial judgment task (line orienta-tion) and a shape judgment task (open vs. closed shape).

.

The investigation of gender differences in global–localprocessing in the context of a spatial and a non-spatial task will al-low us to examine whether these differences, if they exist, charac-terize visual perception of women and men in general, or whetherthey are related only to visuospatial performance.

The main purpose of the present study was to systematicallyexamine gender differences in global–local processing.

The inconsistencies in the results of these studies are mostprobably due to the differences in stimulus and task variables,which are known to affect global–local performance, as discussedearlier.

Direct examinations of gender differences in global–local pro-cessing are sparse, and the results are equivocal.

.

.

The gender differences in navigation and way-finding also ap-pear to suggest a global bias for men versus a local bias for women.

.

.

Gender differences in spatial abilities have been reported in anumber of studies over the years (see Voyer, Voyer, & Bryden,1995, for a review).

Direct examinations of gender differences in global–local processing are sparse, and the results are incon-sistent. We examined this issue with a visuospatial judgment task and with a shape judgment task.Women and men were presented with hierarchical stimuli that varied in closure (open or closed shape)or in line orientation (oblique or horizontal/vertical) at the global or local level. The task was to classifythe stimuli on the basis of the variation at the global level (global classification) or at the local level (localclassification).

This finding suggests that women aremore distracted than men by misleading global oriented context when performing local orientation judg-ments, perhaps because women and men differ in their ability to use cognitive schemes to compensatefor the distracting effects of the global context. Our findings further suggest that whether or not genderdifferences arise depends not only on the nature of the visual task but also on the visual context.

.

Nevertheless, we conducted linear mixedmodels to statistically test for developmental effects to assess whether the magnitude of genderdifferences in confidence increases (or decreases) with grade and found no evidence ofdevelopmental effects in gender differences in confidence (Appendix 4).

Finally, our analyses included data from participants ranging from early childhood toadulthood, but we chose not to focus on developmental trends in our outcomes of interest.

.

.

Across allparticipants, we found the mean values of gamma were positive for both genders (boys/men: n =309, M = .18, SD = .31; girls/women: n = 400, M = .20, SD = .28), which suggests that participantshave some ability to monitor the accuracy of their estimates. However, no reliable gender differencewas observed for relative accuracy (forest plot and analyses are displayed in Appendix 3). Takentogether, the present outcomes suggest that although girls/women (as compared to boys/men) areless confident in their task performance, they are equally able to discriminate between estimates thatare more versus less precise.

Although we cannot calculate measures of absolute metacognitive accuracy, we were able toassess whether gender differences exist for relative accuracy, or the degree to which participants candiscriminate between number-line estimates that are more (vs. less) precise.

Because of these issues, assessing the degree of over- or under-confidence for the number-line estimation task will require future advances in measurement ofjudgments and performance (so they can be made on comparable scales) for this task.

In the current study, we found gender differences in confidence, even when controlling forperformance. We also found that the magnitude of the gender differences observed forconfidence were smaller than those for performance (g = .30 versus .52).

.

.

That is, as compared toboys/men, girls/women may generally have less task-specific efficacy for number-line esti-mation (either because of its math component, spatial component, or both) and may also haveless perceived familiarity with the numbers. In fact, lower self-efficacy may push some girls/women away from engaging in math tasks, which in turn could reduce their actual familiaritywith those numbers.

As mentioned in the Introduction, confidence judgments can be influ-enced by multiple theory- and experience-based factors (e.g., Undorf et al. 2018). The observedgender gap in confidence could be explained by differences in judgment cue use by girls/womenand boys/men.

Although considerable debate exists over the causes of such gender differences when they areobserved (e.g., Hyde 2014), we imagine that psychological (e.g., differences in math attitudes;Sidney et al. 2019), social (e.g., differences in early spatial experiences, such as exposure to spatiallanguage, media, and toys; Caldera et al. 1989; Doyle et al. 2012; Pruden and Levine 2017; orgendered stereotypes about math and spatial ability, McGlone and Aronson 2006; Moè andPazzaglia 2006), and possibly even biological factors (e.g., sexual dimorphism in the parietalcortex; Goldstein et al. 2001) could contribute to the gender differences observed in the number-line estimation task (as is the case for performance; e.g., Tosto et al. 2018).

In summary, the current study indicates that sex differences inglobal self-assessments of performance do not always coincide with sexdifferences in moment-to-moment spatial performance monitoring.Even though female students were in most cases less confident thanmale students in their general spatial ability, their trial-by-trial meta-cognitive monitoring accuracy was not impaired in either an absoluteor relative sense. Thus, female students appear to have relatively ac-curate perceptions of their spatial performance for spatial orientationand spatial visualization tasks.

.

Regardless, the current results show that even female studentspursuing STEM degrees are less confident than male students in theirability to reason spatially about some STEM related content. It is un-clear if these differences reflect true underconfidence or if they are dueto actual differences in academic spatial ability.

This reveals an interesting pattern:for the left ROI, both ipsi- and contralateral attention conditionslead to a decrease in alpha power compared with pre-cue values.However, in the right ROI, we observed a contralateral decreasebut a slight ipsilateral increase.

For the left ROI, the observed decreases were signifi-cant for both conditions (Fig. 6, p
0.05; significant time sam-ples indicated). For the right ROI, only the contralateral attentioncondition lead to a trend ( p  0.064).

A lin-ear regression analysis showed that with decreasing cue reliabil-ity, there was a strong trend toward decreasing differences in thealpha-lateralization index between correct and incorrect trials(R 2  0.043, p  0.081). This effect was significant for reactiontimes: with decreasing cue reliability, the difference in the alpha-lateralization index between low- and high-RT trials becomessmaller and eventually flips from positive to negative values(R 2  0.135, p
0.01)

As expected, in the 50% condition alpha-lateralization index val-ues were rather low and did not correlate with performance (data notshown): both correct and incorrect, and low- and high-RT trialsshowed similarly low alpha-lateralization index values (in the rangeof 0.01– 0.02; correct vs incorrect: t(17)  0.436, p  0.669; low vshigh RT: t(17)  0.375, p  0.712).

For discrimination rate this effect was not sig-nificant (t(17)  0.638, p  0.532), for RT a near significant trend wasobserved (t(17)  2.048, p  0.056).

This was further substantiated by analysis of the invalid cue trialsfrom the 75% condition, on which an opposite pattern was ob-served: high alpha lateralization was detrimental for performance oninvalid cue trials (Fig. 4C,D).

This analysis showsthat a higher alpha-lateralization index precedes better perfor-mance: both correct trials and fast RTs are related to high alphalateralization values whereas incorrect trials and slow RTs showless alpha lateralization. Paired-sample t tests confirmed thatthese differences were significant (correct vs incorrect: t(17) 2.187, p
0.05; low vs high RT: t(17)  2.556, p
0.05).

alpha-lateralization index showed no significant effect (R 2 0.000, p  0.964).

This decrease could not be explained by a difference in overallipsilateral plus contralateral alpha power between the conditions(data not shown), as a similar test on the denominator (normal-ized per subject using the average power over conditions) in the

There was a significant parametric decrease of thealpha-lateralization index with decreasing cue reliability (Fig.3B) as assessed by linear regression (R 2  0.150, p
0.01).

The alpha lateralization was significanton sensor level both in the 100% (see before) and 75% condi-tion ( p
0.01 for two clusters above left and right sensori-motor regions). For the 50% condition the effect was muchweaker, however, a significant cluster was found in sensorsover left sensorimotor regions ( p
0.05).

A cluster-based randomization test overthe 3D source space showed that the lateralized difference inalpha activity between attention-left and attention-right was sig-nificant for the right somatosensory source ( p
0.01) andshowed a trend for the left source ( p  0.062). Note that data ofonly 17 subjects was used for the source analysis.