Human lateral preferences, such as handedness and footedness, have interested researchers for decades due to their pronounced asymmetries at the population level. While there are good estimates on the prevalence of handedness in the population, there is no large-scale estimation on the prevalence of footedness. Furthermore, the relationship between footedness and handedness still remains elusive. Here, we conducted meta-analyses with four different classification systems for footedness on 145,135 individuals across 164 studies including new data from the ALSPAC cohort. The study aimed to determine a reliable point estimate of footedness, to study the association between footedness and handedness, and to investigate moderating factors influencing footedness. We showed that the prevalence of atypical footedness ranges between 12.10% using the most conservative criterion of left-footedness to 23.7% including all left- and mixed-footers as a single non-right category. As many as 60.1% of left-handers were left-footed whereas only 3.2% of right-handers were left-footed. Males were 4.1% more often non-right-footed compared to females. Individuals with psychiatric and neurodevelopmental disorders exhibited a higher prevalence of non-right-footedness. Furthermore, the presence of mixed-footedness was higher in children compared to adults and left-footedness was increased in athletes compared to the general population. Finally, we showed that footedness is only marginally influenced by cultural and social factors, which play a crucial role in the determination of handedness. Overall, this study provides new and useful reference data for laterality research. Furthermore, the data suggest that footedness is a valuable phenotype for the study of lateral motor biases, its underlying genetics and neurodevelopment. Most laterality research in humans has studied the prevalence and origins of handedness as it has been used as a proxy for asymmetrical cerebral organization8. However, humans demonstrate another highly established limb preference, namely to favour one foot over the other10. However, to this day, a precise estimate on the prevalence of atypical footedness in the population is still missingA consistent behavioural phenomenon in many species is that most prefer one limb over the other11 for example assessed language lateralization in 373 participants via a dichotic listening task and found that footedness was the single best predictor. In a large-scale study investigating 1554 individuals, footedness was also a better predictor of language lateralization as compared to handedness on the categorical level12. These results were supported by other studies and could be extended to emotional lateralization as well15. The reason for this increased association with these phenotypes might be linked to the “purity” of the measure. Footedness is most likely less influenced by cultural norms or social teaching compared to handedness for which influences of these factors have been documented repeatedly19. For that reason, footedness might be a better suited candidate phenotype to approximate brain lateralization in large cohorts.Footedness has been suggested to be an on a par or even better predictor than handedness for the lateralization of language in the brain. Searleman20. However, there is much uncertainty whether the extent of left-footedness follows a similar distribution. Illuminating the relationship between handedness and footedness could potentially give insights for the design of studies investigating the genetic basis of motor lateralization. Several studies have investigated the relationship between handedness and footedness and found a significant positive correlation suggesting that there is a shared underlying biological mechanism giving rise to lateralized motor functions26. However, the correlations found in these studies differ vastly ranging from 0.2 to 0.8 indicating that individual studies are unlikely to provide reliable estimates about this relationship in the population.Another important reason to study footedness relates to the precise relationship between different lateral motor biases. It still remains rather unclear to what extent footedness and handedness are associated. Recently, the prevalence of handedness has been investigated in a large-scale meta-analysis and it was identified that 10.60% of the population are left-handed27. If a random sample is drawn from the population, this skewedness then requires a very large sample size in order to represent each group with sufficient power25. Furthermore, due to a large variety of classification systems and assessment batteries, it is difficult to compare results from individual studies even if they comprised large sample sizes. For example, Tran and Voracek28 found that in a substantial sample of 12,720 participants, there were 1,043 (8.2%) left-footers and 3,839 (30.2%) mixed-footers. Martin and Porac29 sampled another large cohort of 3,716 participants and found a substantially larger proportion of left-footed individuals (560 or 15.1%), but almost no mixed-footers (68 or 1.8%). As evident from these results, individual samples can hardly achieve a representative prevalence of the population due to potential sampling biases and different measurements of footedness. Furthermore, both the fields of psychology and neuroscience have suffered from a massive replication crisis due to underpowered studies reporting spurious effects, publication bias, and p-hacking31. One solution to these issues is the application of large-scale meta-analyses as they can provide meaningful insights into the true point estimates in the population, are more resistant to sampling errors, and can identify publication biases. They additionally allow for the investigation of moderating factors that potentially influence foot preference32.A major problem in providing a precise prevalence of atypical footedness and finding a reliable estimate on the association with handedness stems from methodological problems in laterality research. First, lateral motor preferences are skewed within the population with right-sided preferences being more prevalent than left-sided and mixed preferences33, autism spectrum disorders , dyslexia35, intellectual disability36, but also post-traumatic stress disorder (PTSD) and depression38. It is thus conceivable that there is a similar relation with footedness as these lateral biases are significantly associated. A higher prevalence of atypical footedness has been found in individuals with schizophrenia spectrum disorders41. In a large sample of 2,800 participants, it was even shown that mixed-footedness is a better predictor than mixed-handedness for schizotypical traits, potentially due to the reduced cultural influences on the measure26. In neurodevelopmental disorders, such as dyslexia and attention deficit hyperactive disorder (ADHD), footedness has been found to be significantly more atypical, i.e. less right-lateralized, compared to healthy controls44. Studies investigating the relationship between footedness and ASD found no association46. However, these studies were severely underpowered with sample sizes as low as 12 individuals calling into question whether they had the means to identify any effects. In conclusion, there is evidence from individual studies that footedness might be linked to similar clinical disorders as is the case for handedness.There is a wide body of literature claiming that individuals with psychiatric and neurodevelopmental disorders exhibit elevated rates of non-right-handedness, for example in schizophrenia49. Furthermore, players with ambilateral skill are more polyvalent in their position on the field as they can be used on either side. Grouios and colleagues50 investigated the prevalence of mixed-footedness in 1,509 football players and indeed found higher rates of mixed-footedness in professional (45.9%) and semi-professional (30.3%) soccer players compared to amateurs (12.6%) or non-sporting university students (9.1%). Interestingly, a study investigating the salary returns of players in the German Bundesliga in relation to their two-footed ability found substantial evidence of higher premiums if players demonstrated ambilateral qualities, even after controlling for other performance scores51.Footedness also represents an important topic in the context of sports. In symmetrical sports like football, ambilaterality is considered advantageous as there is equal opportunity to use either limb facilitating competitive play at the highest level53. Carey et al.54 investigated both the incidence of left-footedness as well as the relative success rate of left- compared to right-footers in professional football players during the 1998 World Cup in France. They found neither evidence of increased rates of left-footedness compared to the general population, nor that left-footed players were overall more successful in their actions. However, while some studies replicated these findings on the prevalence of left-footedness in football players55, Verbeek and colleagues56 found a ratio of 30.9% left-footers amongst 497 football players. Overall, the findings seem to suggest that mixed-footedness is observed more often in football players compared to the general population whereas findings regarding an increased rate of left-footedness remain rather inconclusive.For handedness, not only ambilaterality, but also a left preference has been proposed to provide an advantage in sports as left-sided athletes are encountered less often during practice and therefore opponents will have less experience in playing against them2. For both handedness and footedness, cross-sectional studies have suggested that right-sided preferences in the population increase with age59. Several studies investigating the prevalence of atypical footedness during childhood have reported increased rates of specifically mixed-footedness compared to adult samples whereas the prevalence of left-footedness seemed to be similar in child and adult samples61. A possible explanation for this phenomenon has been proposed by theories of differential hemispheric aging. The right hemi-aging model assumes that the right hemisphere is prone to a quicker functional decline than the left hemisphere63 possibly leading to increased left-hemispheric dominance as humans grow older. It has to be noted however that this model is mostly supported when comparing younger to older adults leaving open whether this model also applies to individual development during childhood and adolescence. A non-biological explanation for this phenomenon is also possible since the assessment of footedness in children has often been performed using single or very few motor tasks like kicking a ball or hopping on one foot65. Thus, the increased prevalence of mixed-footedness might be a result of less accurate test batteries that do not allow for a more specific determination of footedness as is common in adult footedness research. For handedness, it has been suggested that age effects are rather due to generational effects instead of changes in limb preference across the individual lifetime68. However, as there is little evidence for social pressure against left-foot preference, this type of explanation might be a less likely account for possible age effects on the prevalence of left-footedness.Another important and to this day highly debated question is whether motor asymmetries are present from early gestation and stable across the entire lifetime69. Furthermore, we investigated the relationship between footedness and handedness in the population by specifically analysing those studies that determined both upper and lower limb preferences. We additionally investigated moderating factors on the prevalence of atypical footedness, namely (1) whether or not samples came from a population diagnosed with a psychiatric or neurodevelopmental disorder, (2) whether sampling was conducted in experienced athletes or in a non-sporting population and (3) whether samples consisted of children or adult participants. Furthermore, the test instrument during footedness assessment was used as a moderator since the type of measurement used has been demonstrated to influence the proportion of left-handedness20.The first goal of the present meta-analytic study was to provide an accurate and reliable point estimate of atypical footedness in the population. For that purpose, we conducted four meta-analyses using different classification schemes ranging from a conservative trichotomous left-mixed-right classification to a liberal dichotomous non-right/right classification. We then investigated sex differences in footedness since males have been demonstrated to be more often left-handed compared to females71.This study includes new data on footedness from the Avon Longitudinal Study of Parents and Children (ALSPAC) study, a UK population-based longitudinal cohort. Pregnant women resident in the county of Avon, UK, with expected dates of delivery 1st April 1991 to 31st December 1992 were invited to take part in the study, resulting in 14,062 live births and 13,988 children who were alive at 1 year of ageFootedness was assessed using 4 items  rated on a 3-point scale  in both mothers and children. In children, footedness was assessed at 42 months based on maternal report. In mothers, the same items were assessed using self-report. In order to develop a score where the lowest values indicate left-footedness and the highest values indicate right-footedness, “left”-responses were coded as 1, “either”-responses were coded as 2 and “right”-responses were coded as 3 and a mean value was calculated from the number of valid responses. Finally, a summary variable was created by recoding this variable as follows: children: < 2.00: left, >  = 2.00 and =  < 2.50: mixed, > 2.50: right; mothers: < 1.75: left, >  = 1.75 and =  < 2.50: mixed, > 2.50: right.Similarly, handedness was assessed via maternal report in children at 42 months and self-report in mothers. Handedness assessment was based on six items  and a summary score was obtained: children: < 2.00: left, >  = 2.00 and =  < 2.75: mixed, > 2.75: right; mothers: < 1.50: left, >  = 1.50 and =  < 2.60: mixed, > 2.60: right.https://www.bristol.ac.uk/alspac/researchers/our-data/).Ethical approval for the study was obtained from the ALSPAC Ethics and Law Committee and the Local Research Ethics Committees. Informed consent for the use of data collected via questionnaires and clinics was obtained from participants following the recommendations of the ALSPAC Ethics and Law Committee at the time. Please note that the ALSPAC study website contains details of all the data that is available through a fully searchable data dictionary and variable search tool  . The remainder of 8,528 mothers not suffering from any of these conditions were assigned to the control group .72) when the children were 7.5 years old. The DAWBA is a diagnostic tool based on DSM IV criteria for a number of psychiatric disorders . Overall, 140 children had a diagnosis of ADHD , 221 had a diagnosis of oppositional-conduct disorder ), 217 had a diagnosis of any anxiety disorder , and 36 had a diagnosis of a depressive disorder  and were assigned to the clinical sample. This sample consisted of 485 children  . Due to reports of elevated levels of atypical behavioural asymmetries in dyslexia35, we also included a dyslexia sample. After exclusion of the clinical sample, an assignment of dyslexia was based on a performance IQ of >  = 85 at the age of 8.5, an age-adjusted single word reading score  ≤ − 1 SD for both a 40-word reading test at 7.5 years and a 12-word reading test at 9.5 years73. The dyslexia sample consisted of 268 children . Children free from physical disabilities, psychiatric diagnosis and not included in the dyslexia sample were assigned to the control sample . Footedness data for all subgroups in the ALSPAC cohort are included in Supplementary Table 1. Using individual data points, we calculated Spearman rank correlations between footedness and handedness in all subgroups of the ALSPAC cohort. There was strong evidence in all subgroups that footedness and handedness were significantly correlated . 95% confidence intervals did overlap between all groups, suggesting no group differences in the strength of correlation.Footedness data were available for 9,817 children. Of these, 60 were excluded because of reported physical disabilities since this might influence the foot preference. Psychiatric assessment was conducted using the parent version of the Development and Well-being Assessment . The following procedure was used for the selection and inclusion of research studies into the present meta-analyses: in a first step, the electronic databases PubMed (https://www.ncbi.nlm.nih.gov/pubmed/), Web of Science (https://www.webofknowledge.com), and Google Scholar (https://scholar.google.de/) were searched for the terms “footedness”, “foot use” or “feet use”, “foot preference” or “feet preference”, “foot asymmetry” or “feet asymmetry”, “lower limb preference” or “lower limb asymmetry”. All reference lists of eligible publications were then inspected for other research items associated with footedness. Furthermore, review articles on human footedness were inspected76 to identify articles that could not be found in the above mentioned electronic databases. E-mail requests were sent to corresponding authors if (1) articles could not be retrieved from online databases and (2) if the study clearly stated that participants’ footedness was measured, but did not report footedness in a manner to extract the data properly. Data collection was conducted by JP, JS and GB and concluded in February 2020. Data extraction was conducted by JP and JS. Cohen’s Kappa was used to determine interrater reliability as it can account for chance based agreement77. The interrater reliability between JP and JS was at κ = 0.88 which is considered a strong level of agreement in accordance with McHugh78. Disagreements were resolved by discussion.Study selection and conduction of meta-analyses followed the PRISMA guidelines79, arthrosis80, or trisomy81 were excluded from analysis. We specifically included patients with neurodevelopmental or psychiatric disorders  and their control groups. Control groups were treated as healthy cohorts for all analyses.Participants: Only physically healthy participants were used to control for potential influences of physical disabilities on foot preference. Samples including participants diagnosed with conditions such as stroken = 21 studies), they were excluded from the moderator analysis comparing children and adult cohorts.Age: Participants of all ages were included into the study. Datasets were classified as children cohorts if the participants’ age ranged between 0 and 17 years and classified as adult cohorts if the participants’ age was 18 years or above. If the cohort consisted both of children and adults , the present meta-analysis included studies using Left–Right (L–R), Left-Mixed-Right (L-M-R), and non-Right–Right (nonR–R) classification systems. If studies used an even more precise classification system like strong left- and right-footedness, weak left- and right-footedness, and no preferences, these were converted to L-M-R with weak and strong being classified into one category. For age groups, datasets of single studies were separated if a clear distinction of age between the datasets could be identified. In total, 217 datasets from 164 studies including new data from the ALSPAC cohort totalling Sex: We investigated the moderating effects of sex when data was separately reported for males and females.Handedness: We investigated the moderating effects of handedness when data was separately reported for left- and right-handers .Clinical sample: Datasets were divided into healthy cohorts (198 datasets) including control groups from patient studies and clinical cohorts (13 datasets) including patients suffering from various psychiatric and neurodevelopmental disorders.Sporting ability: Datasets were classified into experienced athletes (32 datasets) and amateurs (179 datasets). To be classified as an experienced athlete, studies either explicitly stated so or were classified by the authors based on at least 4 years of experience in the respective sport.Age group: Datasets were classified into children (59 datasets) and adult cohorts (137 datasets).Mean age: when available, age was included as a continuous meta-regressor.15; 62 datasets), the Lateral Preference Inventory , and kicking a ball (39 datasets) as the first three categories since they were used in the large majority of cases. The fourth category contained other preference inventories used for assessing footedness and was therefore termed “other” (71 datasets).Instrument: The assessment instrument for measuring footedness was classified into four categories and then used as a moderator variable. We chose specifically the Waterloo Footedness Questionnaire , graded on a 3-point scale , or on a 5-point scale .Study location: Location of the study was used as a moderator to identify cultural influences on foot preference. The affiliation of the leading author was used to determine study location unless it was otherwise specified. Locations were grouped into the following regions: Europe (94 datasets), America  (63 datasets), Africa (6 datasets), Asia (40 datasets), and New Zealand and Australia (12 datasets).90, year of publication was also analysed as a categorical variable with the five groupings (a) < 1976, (b) 1976–1985, (c) 1986–1995, (d) 1996–2007, and (e) 2008–2019 in accordance with the handedness meta-analysis by Papadatou-Pastou et al.20.Year of publication: We included the publication year as a continuous meta-regressor. Since very early studies have demonstrated to significantly influence the prevalence of handedness, i.e. studies from before 1940Moderator analyses were mainly conducted using the nonR–R classification system as this analysis contained all datasets since it was the most inclusive . However, we specifically conducted moderator analyses in the L-M-R classification for sporting ability, age group, and instrument as the literature has suggested specific effects for left- and mixed-footedness.91. We performed four separate meta-analyses using random effects models, according to the classification scheme of the studies. For all hypothesized effects, we used an alpha level of 5%. For all exploratory effects, we used a Bonferroni corrected threshold to control for multiple comparisons due to the large number of individual tests performed. The following groups were used for analysis:Non-right-footedness (when assessed as a non-right/right (nonR–R) classification).Left-footedness (when assessed as a left–right (L–R) classification).Left-footedness (when assessed as a left-mixed-right (L-M-R) classification).Mixed-footedness (when assessed as a left-mixed-right (L-M-R) classification).Meta-analysis was performed using the R package robumetaStep 1: Footedness prevalence was calculated by dividing the number of left-, mixed-, or non-right-footed participants by sample size for each dataset. Prevalence is reported with corresponding 95% confidence intervals.Step 2: Meta-regression was performed using correlated robust variance estimation (RVE) models.I2 index reflecting the variance explained by heterogeneity across studies. The I2 index levels can be described as low, moderate, and high, when they are close to 25%, 50%, and 75%, respectively92. Additionally, the Tau2 index was computed to specify if there was variance between studies.Step 3: Each grouping was tested for homogeneity using the t-statistic . Very few studies provided footedness broken down by sex and handedness, thus we could not test for an interaction between these two moderators.Step 4: Sex and handedness influences were investigated for each of the four classification schemes. Footedness prevalence was compared between sexes and between handedness groups using the 93 of the R metafor package94 was used to make the funnel plot symmetrical by omitting and/or adding hypothetical data points due to an asymmetrical distribution. For this analysis, we included only published datasets.Step 5: The funnel plot (funnel function) was visually inspected to identify small study bias. Furthermore, we used Egger’s regression test (regtest function) to provide a qualitative estimate of the asymmetry of the funnel plot. Finally, the trim and fill method (trimfill function)Step 6: Moderator analyses were conducted separately for all moderator variables other than handedness and sex on all data sets using the nonR–R classification. Handedness and sex were excluded here as they only comprised a subsample of the data. Interactions with sex and handedness were tested for all other moderators. As for all other conducted analyses, we applied a conservative Bonferroni correction if the specific effect was not hypothesized.The following analysis steps were carried out for each group:The study was approved by the ALSPAC executive committee at the University of Bristol . No further approval was required as no other new data was generated in our study.k = 217 datasets (from 164 studies including the ALSPAC cohort) were included in this meta-analysis, totalling n = 145,135 individuals. The first aim of the meta-analysis was to provide an accurate estimation on the prevalence of atypical footedness in the population. Simple meta-analysis using robumeta resulted in a point estimate of 23.70% for non-right-footedness with a 95% confidence interval (CI) = 21.40%, 25.90% . Egger’s regression test for funnel plot asymmetry revealed clear evidence for small study bias meaning that small studies showed a larger effect compared to large ones . A visual inspection of the funnel plot revealed the same result  = 2.17, p < 0.05). The same analysis was repeated for datasets containing handedness data to identify the relationship between footedness and handedness. Here, the point estimate was 14.80%  for right-handers, 63.00%  for mixed-handers, and 68.50%  for left-handers  = 9.80, p < 0.001) and left-handers  = 14.90, p < 0.001) as compared to right-handers. There was no evidence for a difference in non-right-footedness prevalence between left- and mixed-handers (t(40) = 1.07, p = 0.289).We were further interested if sex differences between individuals affected the prevalence of atypical footedness and therefore repeated the analysis for datasets from which sex data could be extracted. For females, the point estimate was 21.70% . For males, the point estimate was 26.10%   were included in the analysis, totalling n = 121,302 individuals. Simple meta-analysis using robumeta gave an estimate of the left-footedness (L–R) prevalence of 16.10% with a 95% confidence interval (CI) = 14.40%, 17.70%. Heterogeneity among the datasets was found to be high . Egger’s regression test for funnel plot asymmetry revealed clear evidence for small study bias , as did the visual inspection of the funnel plot  = 2.66, p < 0.01). Regarding handedness, the point estimate was 6.28%  for right-handers, 44.30%  for mixed-handers, and 65.87%  for left-handers  = 7.20, p < 0.001) and left-handers  = 15.82, p < 0.001) as compared to right-handers. There was evidence for a difference in left-footedness (L–R) prevalence between left- and mixed-handers (t(38) = 3.69, p < 0.001).As for non-right-footedness, we repeated the analysis for data broken down by sex and handedness. For females, the point estimate was 14.20% . For males, the point estimate was 19.10%   were included in the analysis, totalling n = 99,490 individuals. Simple meta-analysis using robumeta gave an estimate of a left-footedness (L-M-R) prevalence of 12.10% with a 95% confidence interval (CI) = 10.40%, 13.90%. Heterogeneity among the datasets was found to be high . Egger’s regression test for funnel plot asymmetry revealed clear evidence for small study bias , as did the visual inspection of the funnel plot  = 2.07, p < 0.05). For handedness, the point estimate was 3.23%  for right-handers, 22.60%  for mixed-handers, and 60.01%  for left-handers  = 4.27, p < 0.001) and left-handers  = 20.57, p < 0.001) as compared to right-handers. There was evidence for a difference in left-footedness (L-M-R) prevalence between left- and mixed-handers (t(27) = 6.84, p < 0.001).We again investigated sex effects on the prevalence of left-footedness in the L-M-R classification system. For females, the point estimate was 10.90% . For males, the point estimate was 14.80%   were included in the analysis, totalling n = 99,490 individuals. Simple meta-analysis using robumeta gave an estimate of the mixed-footedness prevalence of 20.20% with a 95% confidence interval (CI) = 17.00%, 23.40%. Heterogeneity among the datasets was found to be high . Egger’s regression test for funnel plot asymmetry revealed clear evidence for small study bias , as did the visual inspection of the funnel plot . An effect of handedness could however be found as the point estimate was 19.70%  for right-handers, 49.20%  for mixed-handers, and 23.90%  for left-handers  = 6.87, p < 0.001) as compared to right-handers, while there was no evidence for a difference between left- and right-handers (p = 0.219). There was evidence for a difference in mixed-footedness prevalence between left- and mixed-handers (t(27) = 5.31, p < 0.001).A final comparison for sex differences revealed that the point estimate for females was 17.60%  and 20.80% for males   = 2.45, p < 0.05).We asked whether participants diagnosed with neurodevelopmental or psychiatric disorders demonstrated higher rates of atypical footedness. Sample type  was extracted from 211 datasets (158 studies). The prevalence of non-right-footedness was found to be 22.90%  in control samples (198 datasets) and 30.50%  in clinical samples (13 datasets) .We investigated whether being an experienced athlete significantly moderated the prevalence of atypical footedness. Here, we first investigated the prevalence using a nonR–R classification. Sample type  was extracted from 211 datasets (158 studies). The prevalence of non-right-footedness was found to be 22.80%  in the general population (179 datasets) and 25.50%  in experienced athletes (32 datasets). There was no evidence that these groups differed from another  was found to be 11.40%  in the general population (91 datasets) and 20.00%  in athletes (12 datasets)  = 2.87, p < 0.01). No difference between experienced athletes and the general population could be detected for mixed-footedness however (p = 0.874).Since sporting ability has been claimed to selectively increase the rate of mixed-footednesst(148) = − 1.99, p < 0.05). There was evidence of a handedness by age group interaction with left-handed adults being more often non-right-footed  = 2.42, p < 0.05).Age group was extracted from 196 datasets 150 studies). The prevalence of non-right-footedness was found to be 27.70%  in children’s samples (59 datasets) and 22.50%  in adult samples (137 datasets)  = 2.31, p < 0.05). For mixed-footedness, the prevalence was found to be 26.40%  in children’s samples and 18.10%  in adult samples  = − 2.36, p < 0.05).Since studies specifically reported increased mixed-footedness in children whereas left-footedness was on the same level as in adult cohorts, we repeated the previous analysis in studies using the L-M-R classification system. Age group (children or adult samples) was extracted from 100 datasets (75 studies) reporting the L-M-R classification. For left-footedness, the prevalence was found to be 9.63%  in children’s samples (32 datasets) and 13.79%  in adult samples (68 datasets) see Fig. . There wsee Fig. . There wAn important variable that could mediate the difference between children and adults is the assessment of footedness. We therefore used the instrument for classification as a moderator for both subgroups (children and adults cohorts) to identify whether one group was specifically affected by the means of determining footedness. We found no differential effects of assessment instrument between the two groups in any classification system .p = 0.182). There was evidence for a handedness by age interaction with left-handers being more often non-right-footed with increasing age  = 2.45, p < 0.05).We then used the mean age as a meta-regressor to determine whether age influenced footedness as a continuous variable. The mean age of participants was reported in 139 datasets (108 studies). Meta-regression revealed no evidence of a moderating effect of mean age on the prevalence of non-right-footedness (p > 0.05).While we found no differences between children and adults regarding the influences of the instrument on prevalence of atypical footedness, this analysis did not provide insights about general effects of the chosen tool for assessing footedness, as potential influences on the prevalence of atypical footedness might simply have been identical in children and adults. We therefore used instrument as a moderator across all datasets. The type of instrument for footedness assessment was extracted from 194 datasets (145 studies). The prevalence of non-right-footedness in these datasets was found to be 21.80%  when assessed using the WFQ (62 datasets), 22.00%  when assessed using the LPI (22 datasets), 18.40%  when assessed as a preference to kick a ball (39 datasets), and 28.30%  when assessed using other preference inventories (71 datasets). There was no evidence in favour of a difference in the prevalence of non-right-footedness between instruments  reporting the L-M-R classification. The prevalence of left-footedness (L-M-R) in these datasets was found to be 11.74%  when assessed using the WFQ (26 datasets), 9.68%  when assessed using the LPI (15 datasets), 12.11%  when assessed as a preference to kick a ball (15 datasets), and 13.19%  when assessed using other preference inventories (43 datasets). There was no evidence for a difference in the prevalence of left-footedness (L-M-R) between instruments .We repeated the analysis also in the L-M-R classification since the striking difference in the prevalence of mixed-footedness between the studies of Tran et al. and Corent(69) = − 2.39, p < 0.05) and kicking a ball  = − 4.24, p < 0.001) as compared to the WFQ. There was no evidence for a difference in the prevalence of mixed-footedness when assessed using other preference inventories compared to the WFQ (p = 0.997).The prevalence of mixed-footedness (L-M-R) in these datasets was found to be 24.60%  when assessed using the WFQ (26 datasets), 14.16%  when assessed using the LPI (15 datasets), 8.09%  when assessed as a preference to kick a ball (15 datasets), and 24.59%  when assessed using other preference inventories (43 datasets) .The number of response options was extracted from 201 datasets (151 studies). The prevalence of non-right-footedness was found to be 22.80%  when participants were given two response options (73 datasets), 24.20%  when participants were given three response options (59 datasets), and 21.60%  when participants were given five response options (69 datasets). There was no evidence for a moderating effect of number of response options on the prevalence of non-right-footedness . However, interaction analysis with sex revealed a lower prevalence of non-right-footedness in female participants in Asian countries  = − 4.53, p < 0.001) compared to female participants in European countries.Study location was extracted from 215 datasets (163 studies). The prevalence of non-right-footedness was 25.00%  when study participants were tested in European countries (94 datasets), 24.10%  when study participants were tested in American countries (63 datasets), 21.90%  when study participants were tested in African countries (6 datasets), 19.20%  when study participants were tested in Asian countries (40 datasets), and 26.70%  when study participants were tested in New Zealand or Australia (12 datasets). There was no evidence for a moderating effect of study location on non-right-footedness . There was, however, evidence in favor of a handedness by publication year interaction with mixed-handers being less often non-right-footed in early published studies  = − 2.15, p < 0.05). When year of publication was used as a categorical variable, the prevalence of non-right-footedness was found to be 22.40%  when studies were published before 1976 (5 datasets), 26.20%  when studies were published between 1976 and 1985 (15 datasets), 33.00%  when studies were published between 1986 and 1995 (19 datasets), 21.40%  when studies were published between 1996 and 2007 (70 datasets), and 21.20%  when studies were published between 2008 and 2019 (102 datasets). There was no evidence for a moderating effect of publication year on non-right-footedness .Year of publication was extracted from 211 datasets (158 studies). Meta-regression revealed no evidence of a moderating effect of publication year on the prevalence of non-right-footedness  could not find evidence of a single gene giving rise to hand use107. Recent studies investigating the molecular genetics of hand preference in over 300.000 individuals have successfully identified genes that play a crucial role in the determination of cellular pathways involved in neurodevelopment108. Nonetheless, there is still a wide gap in knowledge regarding the genetic basis of functional asymmetries. One possible reason for the inconsistencies in findings about the genetic foundation of handedness could be the strong nurture component in the ontogenesis of handedness109. Since our study found only a small moderating effect of culture, footedness seems to be the better candidate phenotype to understand the genetic basis of lateralization in motor behaviour and potentially cerebral asymmetries like processing of language and emotion15.Sex differences in footedness have only rarely been investigated, but individual studies have indicated that males are more likely to exhibit left-footedness compared to females110. In our analysis, we found a significant association between atypical footedness and clinical disorders indicating that handedness is not the only lateral bias that plays a distinguished role in this matter. Rather, it seems that there is a connection between clinical disorders and atypical motor biases in general. For example, the typically found left-sided cradling bias in the population111 is shifted to the right in clinical cohorts114. Our finding regarding the association between clinical conditions and footedness therefore highlights the importance of atypical motor asymmetries in psychological and neurological research. However, it remains elusive why that would be the case. It has been proposed that early-life stress might be the missing link between atypical development of both psychiatric and neurodevelopmental disorders and atypical functional hemispheric asymmetries including motor biases118. Here, early life stress could be a factor disrupting typical brain development ultimately causing altered functional as well as structural hemispheric asymmetries. Since there are currently no empirical studies directly investigating this link, we hope that our present results motivate more studies to look beyond the horizon of handedness as a candidate phenotype to help achieve a thorough understanding of the ontogenesis of psychopathologies.The link between handedness and psychiatric and neurodevelopmental disorders has sparked high interest in the field of psychology and medicine alike resulting in a large variety of studies investigating this relationship20. According to our analysis, this also holds true for footedness indicating that left-sidedness could indeed be advantageous in a variety of sports. Interestingly and contrary to our hypotheses, we could find no evidence that mixed-footedness is encountered more often in experienced athletes as has been suggested by Grouios et al.50. A possible reason for this phenomenon could be that for example professional football players, which comprised the large majority of the athlete data, still might have a clear preference to use one over the other even though they are usually more trained in using both feet119.There is evidence of increased left-handedness in experienced athletes in handedness120. However, their stability across the lifespan remains an open question. For footedness, we found that the prevalence of non-right preference is actually increased in childhood compared to adulthood in line with results from for example Iteya et al.121. Interestingly, we were able to identify that the increase of non-right-footedness is exclusively due to higher rates of mixed-footedness in children as left-footedness was found to be higher in adults compared to children. Thus, our meta-analysis indicates that there is a clear shift towards more pronounced motor biases in either direction in adulthood and that brains become overall more asymmetrical during development. This result opposes predictions of the right hemi-aging model as this model solely predicts a shift to more right-sided motor biases62. An alternative explanation for this finding could be found in associations between motor biases and language lateralization. As language production and comprehension is strongly lateralized towards the left hemisphere whereas the right hemisphere specializes in prosody, intonation, and emotional speech122, the development of language skills throughout childhood and adolescence could significantly influence other cerebral asymmetries such as motor biases. Indeed, Szaflarski et al.123 found that there is an increase in language lateralization with increasing age in children and adolescents supporting this hypothesis.Motor asymmetries seem to be present at a very early life stage, even prenatally in the case of handednessA limitation of the present meta-analysis pertains to the general assessment of footedness in the population as it was almost exclusively assessed by a self-report rather than performing the actual task such as kicking the ball. Opposed to handedness, participants are usually less aware with which foot they prefer to perform a certain task from a footedness questionnaire as these activities are less prevalent in everyday life. While we do not deem it likely, the observed differences between the prevalence of handedness and footedness could at least to some extent relate to this issue.Second, even though we used different classification systems in the meta-analysis, the underlying thresholds to determine for example an individual as left- or mixed-handed are entirely arbitrary and vary significantly from study to study. Unfortunately, this variable could not be integrated as a moderator as most studies did not report these thresholds given that the determination of footedness was largely not in the focus of the study. Given the very large study and sample size of the meta-analysis, effects of different thresholds will however likely be minor to the overall results.A third limitation of this study concerns the analysis of clinical vs. healthy samples. Since studies investigating the prevalence of footedness from a single psychiatric or neurodevelopmental disorder were insufficient to conduct meaningful meta-analyses with, we subsumed all disorders into one clinically relevant category. While we found evidence in favour of the hypothesis that these disorders are associated with atypical footedness laterality as is the case for handedness, it remains unclear whether it is specific to a single disorder or a general phenomenon. We therefore urge researchers to assess this phenotype in future research on atypical lateralization patterns in clinical research to illuminate on this crucial issue.Fourth, the comparison of athletes vs. the general population featured studies investigating football players in the large majority of the cases whereas only small subsets of studies investigated for example dancers, cyclists, or skiers. Thus, our obtained results might be specific for football players.Finally, we were unable to investigate the interaction between sex and handedness as footedness was in the vast majority of cases reported separately for both variables. We therefore hope that researchers upload raw data files into online repositories to allow for more specific evaluations of raw data sets in the context of future meta-analyses.125. While the sheer number of studies included in this meta-analysis already indicates the general interest in footedness across various scientific disciplines, there is still a big gap in knowledge regarding its genetic foundation. We hope that this meta-analysis inspires many fields of research to not only look at handedness, but also at footedness, a phenotype that is similarly easy to assess. Large datasets such as the UK biobank should thus also assess footedness in the future. Only if motor biases are investigated holistically might we one day understand their ontogenesis and their role in brain development.Overall, these results could speak in favour of a new perspective in laterality research. Future studies could investigate footedness as a viable candidate phenotype in the context of development and disorders. In that regard, it might be an even better phenotype compared to handedness, as footedness seems less prone to high variance in datasets due to cultural and social influences. It has to be emphasized however that while footedness might be better than handedness as a proxy for brain asymmetries, motor preferences in general remain rather poor predictors of overall cerebral lateralization given the low correlation between these variablesSupplementary Figure S1Supplementary Figure S2Supplementary Figure S3Supplementary MaterialSupplementary Table 1


RR reduction in incidence of CRC  and 62% RR reduction in mortality of CRC . Subgroup analysis of different interventions, study design, country, sample size, age or sex showed that the incidence and mortality reduction remained consistent, and colonoscopy screening had the same effect on people below and above 50. Our study indicated that colonoscopy could significantly reduce the incidence and mortality of CRC.It is the great priority to detect colorectal cancer (CRC) as early as possible, finally to reduce the incidence and mortality of CRC. However, although colonoscopy is recommended in many consensuses, yet no one systematic review is conducted to figure out how colonoscopy could change the incidence and mortality. In our study, we conducted a comprehensive meta-analysis to evaluate the association between colonoscopy screening and the incidence or mortality of CRC. PubMed, EMBASE, and PMC database were systematically searched from their inception to June 2020. A total of 13 cohort and 16 case-control studies comprising 4,713,778 individuals were obtained in this review. Our results showed that colonoscopy was associated with a 52% Colorectal cancer (CRC), one of the most common malignancies In order to detect CRC earlier, the American Cancer Society recommends screening for colorectal cancer from the age of 45, based on epidemiological data and mathematical models. Both the fecal occult blood test (FOBT), fecal DNA test and colonoscopy are the mainstream detecting methods. Compared with the other two methods, colonoscopy is the gold standard for the diagnosis of CRC, and it could meanwhile provide an opportunity for treatment Although it is recommended in the consensus, colonoscopy screening programs have not been widely implemented in many European countries http://www.crd.york.ac.uk/PROSPERO).This systematic review and meta-analysis was conducted on the basis of the Meta-Analysis of Observational Studies in Epidemiology  and Preferred Reporting Items for Systematic Reviews and Meta-analysis (PRISMA) statement  What's more, we reviewed the references of identified studies for further study. The authors examined the titles and abstracts independently and in duplicate to identify studies that might be eligible and then reviewed the full text to determine trials that met the eligibility criteria.A comprehensive electronic literature search was performed on PubMed, EMBASE and PMC databases from inception through June 2020 with the following terms: “colonoscopy or colonoscopy” AND “Colorectal Neoplasm or Colorectal Tumor or Colorectal Carcinoma or Colorectal Cancer or Intestinal Neoplasm or Intestinal Cancer or Gastrointestinal Neoplasm or Gastrointestinal Cancer” AND “relative risk or relative risks or odds ratio or odds ratios or rate ratio or rate ratios or risk ratio or risk ratios or hazard ratio or hazard ratios or ratio” AND “case-control studies or cohort studies or cohort or case-control”. The detailed search strategy was described in HRs), relative risks (RRs), or odds ratios (ORs) with corresponding 95% confidence intervals (CIs) or sufficient data for their estimation. (4) Cohort (prospective or retrospective), or case-control studies. As for exclusion criteria, they were displayed as follows: (1) colonoscopy was assessed only in patients with premalignant conditions, colorectal adenoma, inflammatory bowel disease (IBD), ulcerative colitis (UC) and Crohn disease (CD); (2) Non-English published studies; and (3) protocol, case reports, comments, reviews, expert opinions, conference abstracts, letters, and animal experiments.Observational studies  were included if they met the following criteria: (1) A general population older than 18 years old who are not diagnosed with CRC. (2) The control group did not receive colonoscopy or other examination methods. (3) CRC incidence or mortality confirmed by pathologic diagnosis, and reported risk estimates, such as hazard ratios . If any disagreement achieved, a third reviewer would join in and reached a consensus.For all included studies, the following information was extracted: first author, publication year, design, period, country, sample size, gender, age, intervention, follow-up duration, comparator, frequency and timing of colonoscopic screening, adjustments or matching factors, numbers of outcomes, and effect estimates. The primary outcome was the incidence or mortality of CRC. The quality of each study was evaluated with the Newcastle-Ottawa Quality Rating Scale (NOS), as one of the most useful scales to evaluate the quality of non-randomized studies . Cochrane's Q statistic (p<0.10 suggesting significant heterogeneity) and the 2I statistic  were adopted to qualitatively and quantitatively evaluate heterogeneity across studies, respectively p ≤ 0.1 p < 0.05 unless otherwise specified.Considering low heterogeneities within and between studies, we used the fixed-effects model Figure  After the deletion of duplicate studies, there are 2,614 records were considered potentially relevant. After reviewing the titles and abstracts, a total of 75 articles were considered relevant. Three studies were found to be eligible for inclusion in the manual search process, 78 records left. 49 citations were further excluded after carefully reading the full text. The reasons for exclusion were as follows: without a comparator (n=11), conference abstracts (n=9), no available data (n=2), comments (n=6), review (n=3) and formerly diagnosed CRC (n=15), protocol (n=3). Finally, twenty-nine articles A total of 3,536 studies were included, as is shown in arator n=, confere respectively. Among the eligible 29 studies, sixteen were case-control studies Table 2 presented the characteristics of interventions. Among the included studies, there are 19 studies adopted colonoscopic screening, while the remaining study was followed by diagnostic. The regions included in the study were as follows: one from Japan, five from Europe, three from Canada and twenty from the USA. Fourteen and nine studies only reported the incidence or mortality of CRC, respectively, and five reported both.NOS scores and detailed characteristics of the 29 records are presented in  Among these 29 eligible studies, the scores of Newcastle-Ottawa quality were ranging from 6 to 8. All studies scored six stars or more. Moreover, most studies were adjusted or matched for the following confounders: age (29/29), sex (29/29) (Table 1).Quality assessment was shown in RR and 95% CI within a fixed-effects model and the values were pooled RR = 0.48, 95% CI = 0.46-0.49, indicating that colonoscopy can reduce the CRC mortality of 52% RR . However, there was high heterogeneity among studies . To explore the source of heterogeneity, sensitivity analysis was carried out , indicating that the studies of Ko et al, Brenner et al., and Schoen et al., RR. Hence, these three articles were excluded and meanwhile, the incidence rate decreased slightly , as well as the heterogeneity decreased  . Sensitivity analysis showed that none of these 16 studies could have a great impact on the pooled RR .As for incidence, a total of 19 studies were calculated the combination of Table 3, we conducted a subgroup analysis of CRC incidence reduction after endoscopic screening based on different interventions, study design, country, sample size, age and sex. We found that colonoscopy could significantly reduce the CRC incidence compared with never-screened  . In the subgroup analysis of the study design, colonoscopy provided protection in both cohort  and case-control studies  . In the region-based grouping analysis, the incidence of CRC decreased in both western  and eastern  . Judging from the results of the sample size, colonoscopy can reduce mortality in ≥1 million ; 10,000-100,000  and less than 10,000  . From the age group, colonoscopy screening provided protection in both 20-50  and ≥50  . Similar results were also shown in sex groups  .As presented in RR reduction in CRC morality after the colonoscopic screening within a fixed-effects model. The pooled RR was 0.38 (95% CI = 0.36-0.40) and the heterogeneity was moderate  . To assess whether anyone study had a dominant effect on the meta-analysis RR, each study was excluded, and we found no study markedly affected the summary estimate or P-value for heterogeneity among the other summary estimates .A total of 14 studies reported a 62% Table 5, we conducted a subgroup analysis of CRC mortality reduction after endoscopic screening based on different interventions, country, sample size, age and sex. We found that colonoscopy screening had a more significant protective effect than never-screened  . In the subgroup analysis of the study design, colonoscopy provided protection in both cohort  and case-control studies  . In terms of country, colonoscopy screening provided protection in both western country  and eastern country  . Judging from the results of the sample size, colonoscopy can reduce mortality in these three groups ; 10,000-100,000: RR = 0.394, 95% CI = 0.368-0.423, p ≤ 0.001, 2I = 58.1%; ≤10,000: RR = 0.409, 95% CI = 0.311-0.537, p ≤ 0.001; 2I = 54.4%)) . Similar results were also shown in the age   and sex groups  .As displayed in Figure 6, Begg's test combined with Egger's test was utilized to evaluate the publication bias. In the pooled analysis of CRC incidence or mortality reduction after endoscopic screening, the p values of Begg's test and the p values of Egger's test were all above 0.05, indicating that there was no obvious bias among these studies.As displayed in RR and 52% RR reduction in CRC mortality  and incidence  after colonoscopic inspection.This meta-analysis set out with the aim of assessing the importance of colonoscopic screening in preventing CRC incidence and related mortality. Of all the 29 studies involving 4,713,778 individuals, our study found a link between colonoscopy and the mortality and incidence of CRC. The outcomes revealed that patients might benefit from 62% RR and 62% RR reduction in CRC incidence  and morality  in western countries, which might be a reference for eastern countries. Further prospective studies from China, Japan and Korea are warranted.As far as we know, this systematic review and meta-analysis might be one of the leading few studies assessing the value of colonoscopy screening in reducing the risk of CRC among healthy individuals. Moreover, we had observed good results among the associations between colonoscopy screening and the mortality and incidence of CRC. Of all the twenty-nine studies enrolled, only one was from the eastern country. For the remaining 28 studies, we found that colonoscopy could achieve 51% RR and 51% RR reduction in CRC mortality  and incidence . We found that colonoscopy screening had a similar protective effect on young people under the age of 50. And we need more data to draw more reliable conclusions.Furthermore, the minimum age of regular colonoscopic screening is 50, recommended by developed countries What's more, direct access colonoscopy service for CRC screening produces a positive financial benefit for patients and local health districts In explaining our findings, attention should be also paid to the following aspects. On the one hand, we did not include randomized controlled trials (RCTs), because it was difficult to conduct RCTs, especially in Japan and Europe, where CRC screening has been introduced into national health programs. What's more, colonoscopy utilization has been on the rise in North America Although sensitivity analyses partially explained heterogeneity, the primary source of the heterogeneity is unclear. It could be potentially generated by the inherent relationship between cancer occurrence and the pattern or frequency of colonoscopic inspection. Despite the use of a fixed-effects model in this analysis, it is noteworthy that estimates with high heterogeneity are vulnerable.The results indicated that colonoscopy could significantly reduce the incidence and mortality of CRC. After subgroup analysis of different interventions, study design, country, sample size, age or sex, the outcomes remained consistent. Usually, the recommended age by developed countries for regular colonoscopic screening is 50. Based on our results, the population aged 20-50 years old could also benefit from colonoscopic screening. Further researches were required to verify our findings.Supplementary tables.Click here for additional data file.


Accumulating evidence shows that cytokines play an important role in the proliferation of prostate cancer. This research is trying to determine that IL-18 -607 C/A polymorphism confers susceptibility to prostate cancer. Meta-analysis was used to collect data. The relevant studies were identified through a comprehensive search from PubMed, Excerpta Medica Database (EMBASE), Web of Science, and Chinese Biomedical Literature Database (CBM) to obtain related studies published up to December 6, 2017. The association between interleukin (IL)-18 -607 C/A polymorphism and prostate cancer risk was assessed by odds ratios (ORs) together with their 95% confidence intervals (CIs). Nine case-control studies from 6 articles were eventually identified. In the overall population, there is a significant association between IL-18 -607 C/A polymorphism and prostate cancer risk in recessive  or dominant  models. In the sub-group analysis according to ethnicity, for Asians, IL-18 -607 C/A polymorphism was significantly associated with prostate cancer in allele contrast , homozygote , recessive , and dominant  models, for Caucasians, IL-18 -607 C/A polymorphism was significantly associated with prostate cancer risk in allele contrast , homozygote  and recessive  models. This meta-analysis has shown that IL-18 -607 C/A polymorphism contributes to a decreased risk of prostate cancer risk in the Asian population but an increased risk in the Caucasian population.  Prostate cancer refers to a kind of epithelial malignant tumor that occurs in the prostate of men. In Europe, it is recognized as the most common solid tumor, with an incidence rate of 214 cases per thousand men . Although epidemiological studies have shown that the incidence of prostate cancer in the Asian population is far lower than that in the United States and European countries, the incidence and mortality rate of this disease have increased rapidly in China . However, the etiology of prostate cancer remains unclear. Studies have revealed several risk factors during recent decades, which been associated with increased risk of prostate cancer . Some people develop the disease while others do not, even when everyone is exposed to the same environment, suggesting that gene polymorphism, such as single-nucleotide polymorphism (SNPs) may contribute to prostate cancer carcinogenesis.Interleukin-18 (IL-18) is a multifunctional cytokine that can induce tumor necrosis factor-α (TNF-α) and interferon-gamma (IFN-g), enhance the cytotoxicity of natural killer (NK) cells and up-regulate the Fas ligand (FasL) expression . There is no doubt that IL-18 plays an important role in antitumor immunity. In addition, IL-18 is an critical genetic factor in the proliferation of prostate cancer. Variations in the DNA sequence of IL-18 gene promoter may influence the production and/or activity of IL-18, which modulates an individual’s susceptibility of prostate cancer. IL-18 expression in prostate cancer is also related to the pathologic stage and the transition from androgen-responsive to hormone-refractory of prostate cancer.The IL-18 gene is located on chromosome 11q22.2-q22.3. Cloning and gene expression analysis suggested that two SNPs of the promoter of IL-18 gene at position -607 (rs 1946518) and -137 (rs 187238) may cause differences in transcription factor binding and have an impact on IL-18 gene activity . The important polymorphism of IL-18, the C/A mutation at the -607 loci, exists in a variety of tumors, has been particularly considered. In addition, the relationship between IL-18 -607 C/A (rs1946518) polymorphism and prostate cancer has been investigated in previous studies. Although, most of the studies showed that there was no significant association of IL-18 -607 C/A polymorphism in the proliferation of prostate cancer in the Asian population, the result still requires further confirmation. Therefore, a meta-analysis of all eligible case-control studies was performed in order to verify the association between IL-18 -607 C/A polymorphism and prostate cancer risk in this article. To the best of our knowledge, this is the first meta-analysis to discuss the relationship between IL-18 -607 C/A and prostate cancer risk.Search strategyA comprehensive search was conducted on the electronic databases PubMed, Excerpta Medica Database (EMBASE), Web of Science, and Chinese Biomedical Literature Database (CBM) to obtain related studies published up to December 6, 2017. The method of collecting data is meta-analysis. The search strategy was based on combinations using the following terms: “IL-18” or “Interleukin-18” in combination with “polymorphism” or “variant” or “SNP” or “mutation” in combination with “prostate cancer” or “prostate carcinoma” for all publications on the association between IL-18 -607 C/A polymorphism and prostate cancer risk. No language or country restriction was applied. Other potentially eligible articles were collected by searching from the reference lists of relevant literature and reviews.Inclusion and exclusion criteriaAll eligible studies were collected in the meta-analysis according the following inclusive criteria: (a) independent case-control or cohort studies; (b) possessing IL-18 -607C/A (rs 1946518) polymorphism; (c) the paper must provide sufficient genotype data of both cases and controls with odds ratio (OR) and 95% confidence interval (95% CI); (d) if multiple publications from the same or overlapping population were available, the most informative one was included. Studies were excluded if one of the following existed: (a) not case-control studies; (b) not offering the source of cases and controls or other essential information; (c) studies that contained overlapping data; (d) reviews and repeated literature were also excluded.Data extractionAll available data from the eligible studies were extracted according to the standard protocols by two investigators independently. A third investigator joined in the decision-making process if there were any disagreements. The standard protocols are as follows: first author’s name, publication year, ethnicity of the population, country of origin, source of control, the number of cases and controls, genotyping methods, allele frequency, genotype distribution both in case and controls, and the results of the Hardy-Weinberg equilibrium (HWE) test. Different ethnicities of the population were categorized as Caucasian and Asian. Statistical analysis2)-test based on Q-statistic and I2. If the heterogeneity is acceptable (P>0.10 or I2 < 50% suggested no obvious heterogeneity), the fixed effect model ) will be adopted; otherwise, the random effect model ) will be used to calculate the pooled ORs. The random effect model is a method for disposing heterogeneous data, but it cannot replace the analysis of the source of heterogeneity. Several factors might induce the heterogeneity, including design scheme, measuring method, age, and ethnicity. STATA software  was used to analyze the data. The possible association between IL-18 -607 C/A polymorphism and prostate cancer risk was evaluated by odds ratio (OR) and 95% confidence interval (95% CI) in the allele contrast (C versus. A), homozygote (CC versus. AA), heterozygote (CA versus. AA), recessive (CC versus. CA/AA), and dominant (CC/CA versus. AA) models. The heterogeneity between the studies was assessed by the chi-squared  and prostate cancer risk are displayed in Sensitivity analysisIndividual studies were consecutively omitted in the sensitivity analysis to detect the influence of each study on the pooled OR. In the meta-analysis, no single study significantly influenced the overall results as indicated by sensitivity analysis. That means the results of the meta-analysis were robust and reliable.Publication biasWe adopted the Begg’s funnel plot and Egger’s test to evaluate the publication bias of articles in this meta-analysis. As shown in Many studies in recent years have indicated that SNPs are the most common sources of human genetic variation, and they are likely to contribute to an individual’s susceptibility to cancer . Interest in the genetic susceptibility to cancers has resulted in growing attention to the study of polymorphism of genes involved in tumorigenesis. To offset the lack of epidemiological studies of IL-18 -607 C/A polymorphism and prostate cancer risk, this meta-analysis can provide further data on the relationship between this gene and prostate cancer risk.In this study, we included data from 9 case-control comparisons with more than 3,000 genotyped prostate cancer patients and controls. According to the study design, 3 studies were conducted in a hospital-based design , and 6 in a population-based design . The overall data suggested that IL-18 -607 C/A is not associated with the prostate cancer risk. However, the stratified analysis by ethnicity yielded different findings. In the Asian population, although all the articles included in the meta-analysis showed that there is no association between IL-18 -607 C/A and prostate cancer risk, after meta-analysis, we found that IL-18 -607 C/A is associated with the prostate cancer risk as a protective factor, and is also associated with the prostate cancer risk in Caucasians correspondingly. In general, IL-18 -607 C/A polymorphism plays different roles in prostate cancer susceptibility among different ethnic subgroups. Combining data from as many studies as possible has the advantage of reducing random error .Heterogeneity may be a potential problem impacting the results of meta-analysis. Significant between-study heterogeneity in the overall analysis existed under almost all the gene models in the study. However, after subgroup analysis by ethnicity, the heterogeneity was removed in almost all of the gene models in the Asian population. There are many factors that could have contributed to the high heterogeneity. Racial differences may be the most important reason for the genetic heterogeneity. As we all know, the relationship between the genotypes distributed in different populations may also depend on the population stratification. Second, the source of control is different in different studies, which may have resulted in heterogeneity in the meta-analysis. Other reasons that may account for the heterogeneity included different genotyping method and publication bias. The currently available data on the association between IL-18 -607 C/A polymorphism and prostate cancer in Caucasians are still very limited, so it should be interpreted with caution. Further studies are required to provide an authentic result.Some limitations of this meta-analysis should be acknowledged. First, only published studies in English were included in the meta-analysis, so publication biases and English language biases were unavoidable. Second, the results are based on unadjusted estimates, and a more precise analysis stratified by age, different living environment, lifestyle factors, such as smoking and drinking, and different grades of prostate cancer could be performed if individual data were available. Third, the result should be cautiously interpreted because controls were not uniformly defined. These studies may have included controls that had different risks for developing prostate cancer in the future. Fourth, only one study included in this meta-analysis analyzed the Caucasian population. Therefore, more studies of the Caucasian population are needed in the future to obtain more precise conclusions about the associations between IL-18 -607 C/A polymorphism and prostate cancer risk. Despite these limitations or disadvantages, the meta-analysis still had some advantages. First, this is the first meta-analysis on the association between IL-18 -607 C/A polymorphism and prostate cancer risk. Second, a systematic review of the association of IL-18 -607 C/A polymorphism and prostate cancer risk is statistically far more powerful than any single study. Third, the sensitivity analysis and publication bias analysis showed the stability and credibility of the meta-analysis, and the process of literature selection, data extraction, and data analysis in the meta-analysis was well designed and conducted.In conclusion, the present meta-analysis provides information that there is no association between IL-18 -607 C/A polymorphism and prostate cancer risk in the overall population, but in the stratified analysis by ethnicity, IL-18 -607 C/A polymorphism was found to be associated with prostate cancer risk as a protect factor in the Asian population and also associated with prostate cancer risk in Caucasians. Larger scale primary studies with the consideration of gene-gene and gene-environment interactions are required to further evaluate the interaction of IL-18 -607 C/A polymorphism with prostate cancer susceptibility.


Background: DNA methylation has been widely assessed as a potential biomarker for the early detection of cervical cancer (CC). Herein, we assessed the associations of SOX1 promoter hypermethylation with squamous intraepithelial lesion and CC.Methods: Published studies and genome-wide methylation datasets were searched from electronic databases (up to April 2019). The associations of SOX1 hypermethylation with high-grade squamous intraepithelial lesion (HSIL) and CC risks were evaluated by odds ratios (ORs) and 95% confidence intervals (CIs). The summary receiver operator characteristic test was used to assess the diagnostic value of the SOX1 promoter hypermethylation of CC and intraepithelial neoplasia type III or worse (CIN3+). Trial sequential analysis (TSA) was performed to evaluate the stability of results and estimate the required information size (RIS).Results: In this meta-analysis of 17 published studies, the SOX1 methylation rates increased among low-grade SIL , HSIL (40.75%), and CC (84.56%) specimens. Compared with control specimens, SOX1 promoter hypermethylation progressively increased the risk of HSIL by 4.20-fold (p < 0.001) and CC by 41.26-fold (p < 0.001). The pooled sensitivity of SOX1 methylation was estimated to be 0.85 (95% CI: 0.81–0.88) in differentiating patients with CC, corresponding to a specificity of 0.72 (95% CI: 0.69–0.75) and an area under the curve (AUC) of 0.93. Furthermore, the pooled sensitivity of SOX1 methylation was estimated to be 0.75 (95% CI: 0.72–0.78) in differentiating patients with CIN3+, corresponding to a specificity of 0.71 (95% CI: 0.69–0.73) and an AUC of 0.84. The pooled results of TCGA and GEO datasets showed that all CpG sites in SOX1 were associated with CC and 16 of 19 CpG sites were associated with HSIL. The results of TSA illustrated that the size was sufficient and significant associations were observed.Conclusion: This meta-analysis indicated that SOX1 promoter hypermethylation might have a potential value in the clinical diagnosis of CC and CIN3+. And, methylation level of SOX1 was significantly higher in the presence of viral infection of high-risk HPV. The prevalence of CIN 2+ was very low in cases without high-risk HPV infection and SOX1 unmethylated events. The prevalence of CIN 2+ were slightly higher when one of the factors was present (HPV infection or SOX1 methylation) but significantly higher when both factors were simultaneously considered   2009 guidelines  and 95% confidence intervals (CIs);matched controls were included;written language used was English or Chinese.Studies meeting the following criteria were excluded:reviews, meta-analysis, meeting abstracts, letters, or case reports;control group and sample sizes were unclear;in vitro or ex vivo experiments of cell lines or animals.All included studies were independently extracted by two authors. The following information of each eligible study was extracted: first author's name, publication year, ethnicity, country, methylation detection methods, materials, source of controls, sample size-involved diseases , clinicopathological features  stage, and lymph node metastasis), and quality of studies.https://www.ncbi.nlm.nih.gov/gds/) by using the following keywords: “cervical cancer” and “methylation” and “Homo sapiens.” Methylation signals of the datasets above were detected by Illumina HumanMethylation 27 or 450 k Beadchip. The level of methylation at each CpG island site is expressed as a beta value, which is the ratio of quantile-normalized methylation intensity to total locus intensity (methylated and unmethylated).Genome-wide methylation profiles of the TCGA CESC project were obtained. Quantitative methylation datasets were initially searched from the GEO database . TSA 0.9 software  was applied to assess statistical significance. In this meta-analysis, the type I error was set as 5%, and the type II error rate was set as 20% with a statistical test power of 80%. The cumulative Z-curve crossed the trial sequential monitoring boundary or the required information size, suggesting that the statistical evidence is firm for this meta-analysis. Otherwise, additional studies are essential to reach a conclusive result. If the Z-curve did not cross any boundary, then no significant association existed. If the Z-curve crossed the traditional boundary and the trial sequential monitoring boundary, then the sample size was large enough and a significant association existed.Trial sequential analysis (TSA) was performed on the The quality of included articles and datasets was independently evaluated by two authors (JH and JYL) according to a predefined system derived from the REMARK  and their 95% CIs were calculated to assess the association of SOX1 promoter hypermethylation with LSIL, HSIL, CC, CIN3+, and HPV. Heterogeneity of eligible studies was determined by the Cochran's Q-test and I2 statistic. The random-effect model was applied to pool the results when significant heterogeneity  was present; otherwise, the fixed-effect model was applied. The bivariate meta-analysis model was performed to estimate the sensitivity, specificity, diagnostic odds ratio (DOR), positive likelihood ratio (PLR), and negative likelihood ratio (NLR) and to generate the summary receiver operator characteristic (SROC) curves. Subgroup analyses and meta-regression were conducted to further explore the possible source of heterogeneity based on ethnicity, source of controls, materials, published year (≥2015 and <2015), and quality of studies. Funnel plots and Egger's test were conducted to evaluate the potential publication bias, and PEgger ≤ 0.05 implied the existence of publication bias , meta-analysis (n = 1), meeting abstracts (n = 4), cell lines (n = 3), and insufficient data (n = 7).According to the definitions of the 2001 Bethesda System, the category of LSIL encompassed cytopathic effects of HPV, cervical intraepithelial neoplasia (CIN) 1, and mild dysplasia. The category of HSIL contained moderate or extensive dysplasia and CIN 2 or 3. The flow diagram for the procedures of eligible studies selection in this meta-analysis is shown in SOX1 promoter methylation and clinicopathological features. Among these 17 articles, 12 used exfoliated cells of cervical specimen to detect SOX1 methylation status, 3  from the LSIL specimens  to the HSIL specimens  and ultimately to the CC specimens .A total of 253 LSIL, 427 HSIL, 434 CC, and 1,029 controls specimens were included in this meta-analysis. As shown in SOX1 promoter methylation on LSIL risk, as shown in SOX1 promoter methylation with LSIL risk. Then, eight studies including 427 patients with HSIL and 896 controls were included to assess the effect of SOX1 promoter methylation on HSIL risk. As shown in SOX1 promoter hypermethylation was associated with a 4.20-fold  increased HSIL risk. This association remained significant in all subgroups, and no significant heterogeneity was found in all comparisons.Seven studies including 253 patients with LSIL and 870 controls were included to assess the effect of SOX1 promoter methylation on CC risk. As shown in SOX1 promoter hypermethylation was associated with a 41.26-fold  increased CC risk. In consideration that mild heterogeneity was observed in the overall comparison (I2= 34%), subgroup analyses  increased risk in HSIL compared with LSIL with a low level of heterogeneity (I2= 0%). Seven studies including 253 patients with LSIL and 233 CC were included to assess the effect of SOX1 promoter hypermethylation in CC compared with LSIL. As shown in SOX1 promoter hypermethylation was associated with a 29.78-fold  increased risk in CC compared with LSIL with a low level of heterogeneity (I2= 0%). As shown in SOX1 promoter hypermethylation was associated with a 10.81-fold  increased risk in CC compared with HSIL with a moderate level of heterogeneity (I2= 53%). In order to seek the potential sources of heterogeneity, a Galbraith plot was further depicted, spotting one outlier as major sources of heterogeneity  increased CIN3+ risk. In consideration that high heterogeneity was observed in the overall comparison (I2= 75%), subgroup analyses , and the NLR was 0.19 (95% CI: 0.15–0.24). Our analysis showed that SOX1 hypermethylation could be a useful biomarker for CC .The pooled accuracies for SOX1 methylation were determined to evaluate their usefulness as a biomarker for screening patients with CIN3+. As shown in SOX1 hypermethylation was 3.16 (95% CI: 2.10–4.78), and the NLR was 0.33 (95% CI: 0.26–0.44). Our analysis showed that SOX1 hypermethylation could be a useful biomarker for CIN3+ .The pooled accuracies for SOX1 were included, and the location of these CpG island probes was depicted as shown in p < 0.05 when the methylation level of CC was compared with that of the controls, supporting the effect of SOX1 promoter hypermethylation in CC  in the promoter region of on in CC . All CpGp < 0.05 and 6 showed significant results with p < 0.001 when the methylation level of HSIL was compared with that of the controls, supporting the effect of SOX1 promoter hypermethylation in HSIL  than in those with low TNM stage. The SOX1 methylation level is significantly higher in liver cancer patients with large tumor size (≥5 cm) than in those with small tumor size (<3 cm)  DNA binding domain. They are involved in the coding of transcription factors during embryonic development and apoptosis  and CC by 41.26-fold (p < 0.001). This epigenetic effect of SOX1 above was further verified by extracting quantitative methylation data from the TCGA and GEO databases. We found that all CpG sites had higher methylation levels in CC than in the controls. And most CpG sites had higher methylation levels in HSIL than in the controls. These results illustrate that SOX1 promoter hypermethylation is correlated with the progression of SILs to cervical carcinoma, which is consistent with previous studies  and high (stage III/IV) stages of CC (SOX1 promoter hypermethylation in the clinicopathological features of CC are needed in the future. Second, the pooled results were only performed in Asians, Caucasians, and Brazilians, in this meta-analysis. Further studies involving Africans are still needed. Third, only full-text articles written in English or Chinese were included. Articles in other languages were excluded because of unreadable contents or insufficient data, thereby causing a selection bias.However, this meta-analysis still has some limitations. First, the role of es of CC . This reSOX1 could be a promising biomarker for screening CC, high-grade squamous intraepithelial lesion, and CIN3+.In conclusion, the pooled results in this meta-analysis illustrated that DNA methylation of All datasets generated for this study are included in the article/H-ZT contributed to the design and final approval of the study. HG contributed to performing the data analyses. JH contributed to interpretation of data, completion of figures and tables, and writing the paper. All authors have read and approved the final manuscript.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.


The estimates were higher for warm season and multiple-day lag but lower for old populations. Results from subgroup analyses also indicate a multiple-day lag trend and bigger significant health effects during longer day intervals. Although characteristics of individual studies added modest heterogeneity to the overall estimates, the results remained robust during further analyses and exhibited no evidence of publication bias. Our systematic review and meta-analysis indicate that short-term ambient level ozone exposure was associated with increased risk of COPD hospitalizations. The significant association with multiple-day lag trend indicates that a multiple-day exposure metric should be considered for establishing ambient ozone quality and exposure standards for improvement of population health. Future investigations and meta-analysis studies should include clinical studies as well as more careful lag selection protocol.Chronic obstructive pulmonary disease (COPD) is the third leading cause of death globally and ozone exposure is a main cause of its disease burden. However, studies on COPD hospitalizations from short-term ambient level ozone exposure have not generated consensus results. To address the knowledge gap, comprehensive and systematic searches in several databases were conducted using specific keywords for publications up to February 14, 2020. Random-effect models were used to derive overall excess risk estimates between short-term ambient-level ozone exposure and COPD hospitalizations. The influence analyses were used to test the robustness of the results. Both meta-regression and subgroup analyses were used to explore the sources of heterogeneity and potential modifying factors. Based on the results from 26 eligible studies, the random-effect model analyses show that a 10 µg/m Chronic obstructive pulmonary disease (COPD), which is characterized by progressive and irreversible airflow limitation and chronic inflammation of the lungs, is the third leading cause of death globally . AccordiMeta-analysis is an important approach to synthesizing evidence across studies. So far, four analyses have been conducted to quantify the association between short-term ambient ozone exposure and COPD hospitalizations ,14,15,16To provide an updated and better understanding of the association between short-term ambient ozone exposure and COPD hospitalizations, a systematic review and meta-analysis were conducted with strict inclusion/exclusion criteria and with attention to the source of effect variations.3”, alone and in combination. After excluding duplications using NoteExpress, the titles and abstracts from the publications were reviewed. The potentially eligible ones were selected and retrieved as full texts. Furthermore, references in previous meta-analysis papers were screened for any pertinent articles that were not previously identified. Finally, the acquired articles were reviewed and filtered according to the inclusion and exclusion criteria.With no start date specified, four literature databases  were searched for publications in English before September 19, 2019. The search was further updated in February 14, 2020. Keywords used included “chronic obstructive pulmonary disease,” “COPD,” “ozone,” and “OSelection of publications was undertaken independently by two of the authors (H.G. and K.W.). Afterwards, the two authors discussed their results and disagreements were settled by consensus. A flow diagram for details of the search process in the meta-analysis is given in The criteria were: (1) original studies; (2) study design: case crossover or time series; (3) reliable exposure measurement: ozone exposure was estimated by research grade methods or national monitoring station, reported with no missing or only a few missing days; (4) COPD definition: the International Classification of Disease (ICD), Ninth or Tenth Revision; (5) clear outcome measurements: studies reported quantitative measures of association, or gave sufficient data to calculate the excess risk (ER) and 95% CI; (6) results evaluated the influence of short-term exposure to ambient ozone.The criteria were: (1) incomplete information; (2) non-observational study design; (3) non-original studies; (4) did not provide calculable or reported ER/relative risk (RR)/odds ratio (OR)/percent change and 95% CI; (5) uncertain diagnosis of COPD, or diagnosis according to pulmonary function test; (6) patients with COPD and asthma, or did not mention whether asthma was excluded.The following data from the eligible publications were extracted: first author, year of publication, study location, study design, study period, assessment of ambient ozone exposure, characteristics of temperature, detailed definition of outcome, data source , pollutant model, confounding factors, and statistical model. Data on effect estimates (single-pollutant and multi-pollutant models) of the association (and their 95% CI) between ambient ozone exposure and COPD hospitalizations were also extracted.The data extraction process was also undertaken independently by two of the authors (H.G. and K.W.) and disagreements were settled by consensus. The characteristics of the included studies ,41,42,43Although researchers have suggested to evaluate the quality of time series or case crossover studies , there w3 increment in ambient maximum 8-h ozone concentration. This metric has been reported as closely associated with respiratory health [3, which would ensure comparable units. The OR or RR was also converted to ER in accordance with a previous study [IQR were the relative risk and exposure metrics used in the original study.Excess risk  was pooled from appropriate studies per 10 µg/my health . Howevery health . Our assus study . BrieflySome studies reported multiple estimates as several lagged days, age groups, seasons, or other factors. Considering the bias that age , temperaI2 test was used to examine the heterogeneity. A funnel plot and an Egger test were used to investigate publication bias.Due to different study designs, methods of analysis, different lag exposures, geographical and population differences, and experiences from previous studies ,44,50, hInfluence analyses were performed using both covariance ratio and hat plots to identify possible influential cases . After ep value of less than 0.05 was considered to indicate statistical significance.All statistical analyses were performed with the R software, version 3.6.1 , using the “meta” “metafor” “dmetar” and “forestplot” packages. A This meta-analysis was conducted in accordance with Preferred Reporting Items for Systematic review and Meta-Analyses (PRISMA) guidelines .The initial literature searches yielded 1998 documents , of which 1976 were excluded based on duplication, title and abstract, and full text review. Therefore, 26 original articles were selected for further analyses.3 increase in maximum 8-h ozone concentration was associated with 0.84%  higher COPD hospitalizations.A total of 15 studies with 20 estimates were used to pool overall ER between short-term ambient level ozone exposure and COPD hospitalizations in all age-range and year-round exposure. As illustrated in Additionally, influence analyses  were usep = 0.336) further suggest that publication bias did not exist in the overall effect estimate. When the influential case was excluded, the results from the funnel plot and the Egger test (p = 0.444) remained robust.The Egger test and funnel plot were used to investigate publication bias. The funnel plot analyses did not reveal asymmetry results . The estβ = −1.52, p = 0.061), which gained significance after the influential case was removed .Meta-regression analyses were used to investigate potential sources of heterogeneity, according to a set of key study characteristics including publication year, study design, study location, definition of outcome, data source and influenza. The results indicate that all these factors contributed to the heterogeneity (data not shown). The results from the study design show modifying effects at the borderline significant level  increase in COPD hospitalizations. The overall effect size detected in our study is relatively smaller than that in previous meta-analysis, which could be due to our more stringent criteria for effect data extraction.These 26 carefully selected publications provided sufficient data for our systematic review and meta-analysis to investigate the relationship between short-term ozone exposure and COPD hospitalizations. Of all the air pollutants, ozone is expected to have the strongest correlation with both temporal and meteorological factors, thus only studies with both variables adjusted were used in our analyses. Considering the confounding effects from age and season, subgroup excess risk was estimated only from all age-range populations and year-round ozone exposure. Our analyses demonstrate that, after short-term exposure, each 10 µg/m3 increase in ozone level increased 0.40% for COPD hospitalizations in the ≥ 65 age group. This heterogeneity could be due to the limited statistical power of the small sample size for our meta-analysis and the relatively low modifying effects. Although insignificant modifying effects were found for different data sources, it should be emphasized that, when examining the association between air pollution exposure and respiratory health effects that require medical attention, it is important to distinguish between hospital admissions and emergency department visits. This is because only a small percentage of respiratory emergency department visitors are admitted to hospitals. Therefore, emergency department visits  may represent less serious but more common outcomes [Meta-regression analyses were used to investigate effect modifiers for overall risk estimates. The analyses show that study designs contributed to borderline significant modifications on health effects from ozone exposure. However, this could be a chance finding, because the differential effects of case crossover design came mainly from two studies. From our analyses, no significant modifying effect was observed for influenza-adjusted status, while Wong et al.  found thoutcomes .Results from our subgroup analyses suggest that age and seasonal variations existed. Consistent with findings from previous studies ,47, stroTo our knowledge, this is the first meta-analysis and systematic review which has addressed the differences by lag selection and co-pollutants. Our co-pollutant subgroup analyses included estimates only from the all-age-range populations and from year-round ozone exposure, to avoid further confounders. Although the numbers of estimates were not large, our results demonstrate that short-term ozone exposure was positively associated with COPD hospitalizations, using co-pollutants-adjusted data. Our study also identified meaningful results from different lag subgroups. Many previous studies reported have observations for the lag with the most statistically significant results, instead of all lag structures, but this would bias estimates upward. However, if associations were observed in a multiple-day lag selection but not in a single-day lag selection, single-day lag investigations may underestimate the effects, which is well demonstrated in our results. Indeed, the single-day lag results generally show insignificant and lower pooled results, while the multiple-day lag results show statistical significance with bigger effect estimates during longer day intervals.Current short-term ambient ozone quality standards for most countries are solely based on daily maximum 8-h exposure concentrations ; our resSeveral limitations should be considered here. In most of the selected studies, confounding factors were adjusted differently to each other, which might lead to bias. However, as all the selected studies were controlled for the crucial confounders, for example, temperature, relative humidity, day of week, and long-term factors for the time-series, the limitations would not cause a significant impact on our results. Additionally, the subgroup results are based on relatively small numbers of selected studies. More studies are needed to verify our results.In summary, our results show that short-term ambient ozone exposure was positively associated with COPD hospitalizations and no publication bias was found. The subgroup analyses indicate that the relationship might vary among different age ranges and seasons. The multiple-day lag trend, with more significant effect estimates during longer day intervals, indicates that a multiple day exposure metric needs to be considered for establishing ambient ozone quality standards. However, this lag trend still needs further investigation. Future environmental studies and meta-analyses, which investigate the health effects of short-term ozone exposure, should also consider a lag selection protocol.