Non-melanoma skin cancers (NMSCs) are the most common cancers among fair-skinned people. It is estimated that 2–3 million new cases of NMSCs are diagnosed globally each year. The risk of development increases with age, just like in the case of Parkinson Disease. Due to the general aging of the population and substantially high medical costs of NMSC therapy, NMSCs and Parkinson’s disease (PD) are becoming an increasing health problem. In this paper, we conduct a meta-analysis and systematic review to evaluate the NMSC risk among patients with PD. This study is the first to focus on the effect of different epidemiologic aspects of NMSCs and PD in detail.p < 0.0001). Among all NMSCs, the risk of developing basal cell carcinoma in PD patients was significantly higher : 1.15–1.47; p < 0.0001), contrary to squamous cell carcinoma. Further analysis revealed a significantly higher risk of developing NMSCs in patients with previously diagnosed PD . Our data suggest the necessity for regular skin examination of PD patients, though further studies are required to explore the mechanisms forming this relationship.Patients with Parkinson’s disease (PD) have an increased risk of melanoma compared with the general population. Considering that Nonmelanoma Skin Cancers (NMSCs) share similar risk factors with melanoma, there is a need to understand a possible connection between PD and NMSCs. The aim of the study was the evaluation of NMSC risk among PD patients via meta-analysis and systematic review. A comprehensive search of PubMed, Scopus, and Web of Science databases was conducted, including studies from January 2000 to April 2020. We identified 16 eligible studies including 140291 PD patients. Upon statistical analysis, a significantly higher risk of developing NMSCs in PD patients was found compared with the control group (odds ratio (OR) = 1.25, 95% CI: 1.17–1.33;  Parkinson’s disease (PD) is a neurodegenerative disease which mainly affects people aged 65 or older ,2. For mWe found 364 studies in total. After applying the necessary criteria, the remaining 16 studies were included in the meta-analysis, all of them published between 2002 and 2020. The study selection process is presented in the flow diagram of paper’s selection p = 0.0633; p < 0.0001; Of the 16 studies included in the analysis, 20 different odds ratio (OR) assessment results were obtained for NMSCs in PD patients. When all available studies were included in the analysis, a marginally higher risk of NMSCs was found in patients with PD : 0.99–1.64; In the subgroup analysis by NMSC subtype, we included five studies that provided data separately for BCC and SCC.p < 0.0001; A significantly higher risk of BCC was found in patients with PD  was found in patients with PD . In comparison to the previous meta-analyses ,31, in tp < 0.0001), which is consistent with previously reported data [Our research showed a significantly higher risk of developing NMSCs in patients with PD compared with the control group  and a marginally higher risk of SCC . This difference may result from the fact that not all studies presented data separately for BCC and SCC [p = 0.9518), contrary to the group of patients where NMSCs followed the diagnosis of PD, in which case the risk was significantly higher . This confirms the hypothesis that PD patients may be more prone to the development of sun-induced skin lesions. Therefore, the onset of NMSC after diagnosis of PD could result from disease-specific susceptibility or the photocarcinogenic effect of anti-parkinsonian drugs [Interestingly, our meta-analysis revealed that patients with PD have a higher risk of developing BCC  and a marginally lower risk in female patients. The results are intriguing since NMSCs are more common in males, which may be explained partly by more extensive sun exposure due to working outdoors more frequently [p = 0.0015). Also, analysis of retrospective studies showed a significant effect on the risk of developing NMSCs in patients with PD . The probable difference between ORs may be explained by the fact that the retrospective studies based on medical records of patients could not have been accurately followed up.In our study, sex-specific analysis revealed a significantly lower risk of NMSC in male patients with PD  terms, such as: “PD,” “Carcinoma, Squamous Cell,” “Carcinoma, Basal Cell”, “skin neoplasm”, and text terms, including: “Parkinson disease*”, “non-melanoma skin cancer*”, “basal cell carcinoma*”, “squamous cell carcinoma*”, “skin cancer”, and “keratinocyte carcinoma*”. The search was optimized for each database, taking into consideration differences in search syntaxes. The search was restricted to studies involving humans and published in English.Studies were evaluated independently by three co-authors. The potential studies were evaluated by screening titles and abstracts. The definitive inclusion of studies was made after reviewing the full-text articles. Studies were included in the analysis if they reported a number of patients diagnosed with both NMSCs, PD and risk estimates. The exclusion criteria were as follows: reviews, case reports, meeting abstracts, comments, letters to editor and studies with insufficient data for the meta-analysis.Extracted data from eligible studies included: name of first author; date of the study; the sample of the studied population; the patients’ characteristics, including age, sex, and country; diagnostic criteria for PD and NMSCs; subtype of non-melanoma skin cancer, type of research and risk estimates, including: odds radio (OR), hazard ratio (HR), or standardized incidence ratio (SIR), with 95% confidence intervals or data that enabled their calculation.p < 0.05 or I2 > 50%) and the withdrawal of data from the publication illustrated by funnel plot charts, we performed a further selection of studies. Sensitivity analysis was performed by removing one study at a time to determine if the results would affect a single study, especially in the face of suspicious results or significant heterogeneity. Subgroup analysis was performed by NMSC subtype (BCC or SCC), diagnosis order (NMSCs before/after PD), sex, and study type (prospective or retrospective). A two-sided p-value < 0.05 was considered statistically significant.Data analysis, including comparison of the occurrence and OR of NMSC in patients with PD, was performed with the use of the Statistica software (v. 13 PL). OR and the corresponding 95% confidence intervals (CI) were calculated in patients assigned to the NMSC group, compared with patients assigned to the non-NMSCs group and taking into account the presence or absence of PD in the same study. OR > 1.0 indicates a higher risk or more frequent occurrence of NMSCs in patients with PD compared with controls. Random models were used to calculate the OR. The selection of studies included in subsequent stages of the analysis took into account data on their heterogeneity. The Q test was used to assess heterogeneity, while I2 was used to assess inconsistency. We evaluated the potential bias of publications using Begg’s and Egger’s tests and funnel plot charts. When significant heterogeneity was observed in the analysis (To sum up, this study was the first to focus in a comprehensive manner on different epidemiologic aspects of NMSCs and PD, including subtype of cancer, sex, the temporal relationship between both diseases and study type. We found a higher risk of NMSCs in PD patients. Our data suggest the necessity for regular skin examinations of PD patients. Further studies are required to evaluate the connection between those two diseases.


Previously reported associations between second-to-fourth digit length ratio (2D:4D), a proxy for prenatal androgen load, and transgender identity have been inconsistent. The objectives of the present study were to provide additional original data and an updated meta-analysis concerning this association. In a study of 464 participants, we compared the 2D:4D of transgender individuals with age- and sex-matched controls. Patients were recruited at a specialized psychiatrist’s medical office, whereas controls were hired via flyers, advertisements, and as convenience sample. A random-effects meta-analysis of the literature  also quantifies the overall magnitude of the difference in 2D:4D between transgender individuals and controls. In our study providing new original data, we found a significantly higher (i.e. feminized) left-hand 2D:4D in the male-to-female transgender (MtF) identity [mean age: 32.3 ] than in the male control group [mean age: 34.5 ] with a Cohen’s d = 0.271. Concordantly, the meta-analytic results suggest a significant difference in 2D:4D among MtF individuals compared to male controls , which was even more pronounced when individuals had been diagnosed by a clinician instead of self-identified as transgender . In both studies, no significant results were revealed for female-to-male transgender individuals [mean age: 26.1 ] versus female controls [mean age: 27.2 ]. This original investigation and the updated meta-analysis clarify the association between transgender identity and 2D:4D indicating the influence of prenatal androgen on the development of gender identity in subjects born as males. Evidence suggests that the prevalence has increased over the last decades to up to 5–14 male-to-female transgender (MtF) individuals per 1000 adult males and 2–3 female-to-male transgender (FtM) individuals per 1000 adult females2. This growth in prevalence is possibly due to greater social acceptance, de-pathologization, and greater awareness of therapeutic options2. Despite the growing amount of research in this area, causal mechanisms are still unclear. Animal studies show an association between perinatal testosterone and the size of the bed nucleus of the stria terminalis (BNST)3, a region linked to gender identity in humans4. The theory of early organizational testosterone effects on gender identity is further supported by review evidence that transgender individuals’ brains show changes away from their natal sex and toward their perceived gender5. Additionally, studies examining subjects with congenital adrenal hyperplasia (CAH) or subjects with partial or complete androgen insensitivity syndrome (AIS/CAIS) suggest that (1) levels of gender dysphoria are higher when brain androgenization mismatches gender of upbringing and (2) that levels of gender dysphoria are lower when brain androgenization matches gender of upbringing6. For a more detailed description of gender dysphoria in patients with disorders of sex development and possible confounding variables, see Sadr et al.6. In humans, however, it is not possible to directly investigate the effects of prenatal androgens owing to ethical reasons and the long time period between the intrauterine window and the establishment of gender identity. Thus, this research area relies on proxies such as the second-to-fourth digit length ratio (2D:4D), a biomarker determined by the balance of prenatal testosterone to estrogen7, and therefore indicating the prenatal androgen load8. A lower 2D:4D reflects higher prenatal androgen exposure9 and hence, the 2D:4D is lower in males than in females11. It is frequently used to examine the link between prenatal androgen exposure and postnatal phenotypes including diseases, mental disorders, and behaviors reflecting gender bias or being influenced by first trimester organizational changes induced by testosterone/estrogen ratios12, such as alcohol addiction13, autism spectrum disorders15, sexual orientation16, the risk for different forms of cancer17, and physical prowess18. Concerning gender dysphoria and 2D:4D, contradictory evidence exists, which has been summarized in a meta-analytical commentary conducted in 201719 where the authors report a small effect size  indicating feminized 2D:4D of the right hand (but not the left hand) in MtF individuals and no associations at all in FtM individuals. This short systematic review included only few studies with rather small sample sizes or data retrieved from abstracts (which can be influenced by reporting bias), and it did not provide extensive meta-regression and subgroup analyses or distinguish between case–control and cohort studies. In 2020, a more recent meta-analysis6 reported significantly feminized 2D:4D in MtF individuals (d = 0.24). This analysis applied rather strict inclusion criteria (e.g. excluding child samples) and did also not distinguish between patient and cohort samples. The aim of the present research was to re-address this question by adding novel original data from a larger sample size and to subject the entire body of literature to a meta-analysis while taking potentially confounding factors into account. In previous meta-analytic research on 2D:4D and addiction, we found that the effects are more pronounced when cases have been defined via clinical diagnosis instead of via self-disclosure13. Therefore, we especially focused on the difference between transgender individuals diagnosed by a specialized clinician according to classification systems versus self-identification as transgender or gender nonconforming. Gender nonconformity extends the term of transgender individuals insofar that it includes other gender-variant identities or non-binary people, as well. For both the original research and the meta-analysis, we tested the hypothesis that 2D:4D is larger, i.e. feminized, in MtF individuals than in male controls. When comparing FtM individuals with female controls, we decided to follow an exploratory approach since previous  evidence19 is not conclusive. Following the results of our own previous meta-analytic research on 2D:4D13, we defined a second hypothesis regarding the meta-analysis, namely, that the difference between transgender individuals and controls is more pronounced when the transgender person has been diagnosed by a specialized clinician.Transgender identity or gender dysphoria are defined as experiencing an inconsistency between physical phenotype and one’s perceived genderWe wanted to investigate the difference in the second-to-fourth digit length ratio between transgender individuals and control subjects. Therefore, we conducted an original case–control study with large sample size and subjected the entire body of literature to a meta-analysis.27, a line of independent studies aiming to elucidate the role of prenatal sex hormone organization in the etiopathogenesis of psychiatric disorders in childhood, adolescence, and adulthood, such as addictive disorders30, suicide completion31, or behavioral symptoms in children32. Another goal was to establish prevention targets and strategies based on this model33.This study investigated a possible association of 2D:4D and transsexuality in a large sample. It was part of the Finger Length in Psychiatry (FLIP) projecttranssexualism according to ICD-1034 and gender dysphoria according to DSM-535 and 101 male and 102 female age-matched control subjects were analyzed. The patients were enrolled in the study by a psychiatrist specialized in transsexualism (T.M.) at a medical office in Nuremberg, Germany. The control subjects were hired via flyers and advertisements  or as convenience sample . They were recruited by a medical student (I.D.) in the Department of Psychiatry and Psychotherapy of the University Hospital Erlangen, Germany.From 2016 to 2019, we recruited MtF and FtM patients and male and female control groups. In total, datasets of 110 MtF and 151 FtM patients diagnosed with 2. Exclusion criteria consisted of severe physical illness, the diagnosis of an androgen insensitivity syndrome or an adrenogenital syndrome, acute suicidality, pregnancy, breastfeeding, and the endocrine disorders Morbus Addison, Morbus Cushing, diabetes type 1 and type 2, and endocrine cancers. We did also not enroll control subjects taking regular medications (except for contraceptives), with any history of psychiatric in-patient treatment, or an ambulatory consultation due to a mental illness during the preceding 10 years.Inclusion criteria for both groups consisted of providing written informed consent, age ≥ 18 years, and body mass index (BMI) of 18.5–35.0 kg/m36 and the 24 items of the Utrecht Gender Dysphoria Scale (UGDS)37 into German. Afterwards, discrepancies were discussed, and a single translation was selected in agreement. Whereas the original GIDYQ-AA uses a mean with low scores implying gender dysphoria, the original UGDS uses a sum score with high scores implying gender dysphoria. In order to increase comparability, we reversed the GIDYQ scores and converted item scores in the UGDS to mean values. Thus, higher scores of the gender dysphoria scales used in this study represent higher gender dysphoria. The minimum and maximum scores possible were 1 and 5 . High internal consistencies  were found for the German versions: (1) 0.984 for the male GIDYQ (n = 197); (2) 0.985 for the female GIDYQ (n = 238); (3) 0.989 for the male UGDS (n = 197); and (4) 0.981 for the female UGDS (n = 238). Separate Cronbach’s alpha values for transgender individuals and controls are provided in Supplementary Table p < 0.001) for the MtF sample (n = 96), r = 0.087 (p = 0.389) for the male sample (n = 101), r = 0.240 (p = 0.005) for the FtM sample (n = 136), and r = 0.474 (p < 0.001) for the female sample (n = 102).Independent from each other, three study team members  translated the 54 items of the Gender Identity/Gender Dysphoria Questionnaire for Adolescents and Adults 38 we examined early vs. late onset gender dysphoria in transgender individuals by asking about the onset time point (before vs. after puberty) and about the subjects’ discomfort with the assigned gender at birth during their childhood (present vs. absent).We assessed the participants’ sexual orientation on a scale from 1 (= attracted only to women) to 7 (= attracted only to men). Following a classification proposed by Lawrence et al.https://www.gimp.org). Each finger was measured three times by each of three independent raters . The raters evaluated the scans of patients and controls separately. The files were numbered consecutively (controls) or with a patient’s code. Thus, the raters were blind concerning the sex of the control or the patients’ status as MtF or FtM individual. 2D:4D values were calculated for the mean of right-hand and left-hand 2D:4D (M2D:4D), right-hand 2D:4D (R2D:4D), left-hand 2D:4D (L2D:4D), and differences between R2D:4D and L2D:4D (2D:4Dr-l) since low 2D:4Dr-l values have been associated with high prenatal testosterone load, as well8. We found high interrater reliabilities : (1) M2D:4D: n = 462, 0.990; (2) R2D:4D: n = 464, 0.986; (3) L2D:4D: n = 462, 0.983; and (4) 2D:4Dr-l: n = 462, 0.946.Right and left hands of the participants were scanned on common document scanners. The lengths of the index and ring fingers  were quantified using the GNU Image Manipulation Program . All participants provided written informed consent.t-test. Levene's test for homogeneity of variance was also used and if necessary, the degrees of freedom were adjusted. The χ2 test was used to evaluate differences in the frequency of nominal variables. Pearson’s correlations were used to investigate continuous relationships between the 2D:4D measures and GIDYQ and UGDS scores in each group. We present the data as means and standard deviations or relative frequencies. Statistical significance was determined at p < 0.05 (2-sided). The data were analyzed using IBM SPSS statistics Version 24 for Windows .For each natal sex individually, we compared M2D:4D, R2D:4D, L2D:4D, and 2D:4Dr-l between the patient and the age-matched control groups using the Student’s The MtF group did not significantly differ from the male group with regard to age, months of employment during the previous year, BMI, or marital status Table . We founWe found higher M2D:4D, R2D:4D, and L2D:4D in MtF than in male participants and lower M2D:4D, R2D:4D, and L2D:4D in FtM than in female participants; however, only the higher L2D:4D in MtF than in male participants reached significance with an effect size of Cohen’s d = 0.271 Table . Our ana19 underlying their meta-analytical effect size suggests that a sample size of at least 872 participants would be necessary to be able to detect small effects. Due to small prevalence rates of transgender identity, it is difficult to reach such numbers in clinical settings.We found significantly feminized 2D:4D in MtF individuals compared to male controls in the left hand, but not in the right hand and no significant results in FtM individuals versus female controls. However, the nonsignificant differences were in accordance with the prenatal androgen hypothesis. Comprising 464 participants, to our knowledge, this investigation presents the largest sample size among case–control studies of 2D:4D and transgender identity. Nevertheless, a power analysis by Voracek et al.20). Moreover, the low standard deviations (SDs) in both the GIDYQ and the UGDS could attenuate any association with 2D:4D, thus entailing non-significant results. The low SDs further bring small Cronbach’s alpha values  statement46 on the basis of a full-text review.A two-step literature search was conducted using Google Scholar and PubMed and including abstracts in English from study inception until December 10, 2018. In the course of the review process we updated this literature search from study inception until February 28, 2020. The following search terms were combined in several ways: 13 and are detailed in our coding protocol  and one research assistant (C.R.), independently of one another. Disagreement was resolved by discussion and agreeing upon the extracted values. Regarding quality assessment, the extractors’ values were averaged. Following our approach in Siegmann et al.50 within the open-source software environment R, version 3.4.251.All analyses were conducted and all figures were made using the metafor package52, (3) people identifying as tomboys, and (4) studies reporting correlative data. Correlative data were transformed into Hedges’ g using common transformation formulas53.We estimated the standardized mean difference (Hedges’ g) in 2D:4D among MtF individuals and male controls in addition to among FtM individuals and female controls. In this analysis, a more feminized 2D:4D among MtF individuals compared to male controls is reflected by a positive Hedges’ g, whereas a more masculinized 2D:4D among FtM individuals compared to female controls is reflected by a negative Hedges’ g. Furthermore, we tested whether these mean differences were more pronounced among transgender individuals who had been reliably diagnosed by a clinician. For this purpose, we split the sample into transgender patients diagnosed by a clinician using common classification systems , ICD-10) versus transgender patients not diagnosed by a clinician. The group of transgender individuals not diagnosed by a clinician consisted of (1) people subthreshold for the diagnosis, (2) people identifying as gender-variant, assessed with a shortened adaptation of the cross gender questionnaire8. We computed 2D:4Dr-l as the difference between the mean R2D:4D and mean L2D:4D, and related standard deviations were approximated by the pooled standard deviation of the R2D:4D and L2D:4D variances.Moreover, we estimated Hedges’ g in the difference of R2D:4D and L2D:4D (2D:4Dr-l) among transgender individuals and controls since low 2D:4Dr-l values have been associated with high prenatal testosterone load54, dominant hand only42) entailing that our reported meta-analytic 2D:4D effect size refers to an aggregated 2D:4D rather than to M2D:4D. 2D:4Dr-l did not flow into this outcome measure. Heterogeneity among effect sizes within datasets was assessed using the I2 statistic.We performed univariate random-effects meta-analyses using restricted maximum likelihood estimations in which the point estimate for each study was weighted by the inverse of its variance. Non-independence among effect sizes was accounted for by aggregating, for example, in cases in which studies reported separate data for R2D:4D and L2D:4D. Two studies reported solely one 2D:4D value , mean age of participants, and procedure of measuring 2D:4D . Thereby, the slope of the meta-regression line (β coefficient) indicated the strength of the association between the moderator and outcome. All meta-regressions were Bonferroni-corrected for multiple testing.56. Recent meta-analyses on 2D:4D and gender identity19 report descriptive differences (insofar that the association is stronger with R2D:4D), but did not test for significant differences in these subgroups. We ran one post-hoc subgroup analysis examining differences between studies where the 2D:4D raters have been blinded vs. not blinded to the participants’ gender identity.Moreover, we performed prespecified subgroup analyses to investigate the difference in the outcome measures between the right and the left hand, the different methods of measuring 2D:4D  and the classification systems . A comparison of DSM-5 versus ICD-10 was not possible due to small group sizes. “Measurement without soft-tissue deformation” comprised X-rays and direct measurement from the participants’ palm, and “measurement with soft-tissue deformation” comprised photocopies and hand scans. It is a prominent finding in 2D:4D research that associations with target traits are more pronounced in the right than in the left hand e.g.57. Following the authors’ original proposition, we considered analyses to be biased if the intercept differed from zero at p = 0.1057. We evaluated the sensitivity of our analysis by comparing models with and without effects that we assume to be influential outliers58. On the one hand, we defined statistically influential outliers following the approach by Viechtbauer and Cheung58: a study may be considered to be influential if at least one of the following is true: (1) the absolute DFFITS value is larger than 3√(p/(k − p)), where p is the number of model coefficients, and k the number of studies. (2) The lower tail area of a Chi-squared distribution with p degrees of freedom cut off by the Cook’s distance is larger than 50%. (3) The hat value is larger than 3(p/k). (4) Any DFBETAS value is larger than 158. On the other hand, we considered studies to be influential outliers when the sample was substantially divergent to all other samples included. This is the case for one study with two child samples59. The literature is controversial as to whether symptoms of gender identity disorder in children persist into adulthood61, indicating that these samples potentially deviate from the rest of the studies included in our analysis.Small study effects were assessed by visual detection of asymmetries in contour-enhanced funnel plots and by the Egger regression testp < 0.05 (2-sided) was considered statistically significant, except for the Egger test57 as stated above.66 comprising 16 independent samples. Additionally, we conducted one investigation by ourselves described in the first part of this manuscript . This results in 13 eligible articles added up to 17 independent samples. The characteristics of all included studies are detailed in the Supplementary Table The literature search is summarized in the PRISMA flow chart , ICD-10, or DSM-5, and between studies where 2D:4D raters have been blinded or not to the subjects’ case–control status. The exact figures are detailed in the Supplementary Table None of the meta-regression analyses concerning study quality, mean age, and the procedure of measuring 2D:4D showed any significant effects with a Bonferroni-corrected threshold of p = 0.725) and funnel plot  regarding FtM individuals vs. female controls, however, revealed asymmetry in the published literature. Nevertheless, the applied trim and fill method67 suggested to compensate for this possible bias by adding 0 studies (SE = 2.571). Sensitivity analyses revealed no statistically influential outliers. When excluding one study examining two samples of children59, the effect size for FtM individuals versus female controls did not change substantially, whereas the effect size for MtF individuals versus male controls increased .Concerning MtF individuals vs. male controls, the Egger test  but high for the FtM versus female comparison (I2 = 89.1%), possibly resulting in slightly overestimated effect sizes in the latter sample. The high heterogeneity could not be explained by the pre-specified moderators since none of the meta-regression and subgroup analyses revealed any significant results. This contradicts the prominent 2D:4D findings concerning the influence of the measurement with or without soft tissue deformation on the relationship between 2D:4D and target traits11 and of more pronounced effects in the right compared to the left hand due to possible increased sensitivity of the right hand to prenatal testosterone exposure68. We also did not find a significant association between 2D:4Dr-l and transgender identity, which suggests that 2D:4Dr-l is not as suitable as 2D:4D to indicate relationships in this research area. Sensitivity analysis showed that the influence of single studies on the overall effect should be estimated minor.In accordance with our hypotheses, we found associations of small effect size between feminized 2D:4D and MtF gender identity that were more pronounced in MtF individuals diagnosed by a specialized clinician. No significant results have been revealed in FtM individuals compared to female controls. Between-study heterogeneity was small for the MtF versus male sample . However, with a non-overlap of 14.7% between transgender individuals and controls, one has to bear in mind that the practical impact is small, too69.Both the original study and the meta-analysis support our hypothesis that MtF individuals have a higher (i.e. feminized) 2D:4D than male controls. The effect was amplified when diagnoses were made according to classification criteria of gender dysphoria or transsexualism instead of self-identifying as transgender. This indicates that in males a lower prenatal androgen load associates with more pronounced gender dysphoria. In terms of practical relevance, the small effect sizes of up to Hedges’ g = 0.193 mean that the 2D:4D of 58% of male controls is below the average 2D:4D of MtF individuals70, which can result in a discrepancy of sex between body and brain71. The lack of the second testosterone secretion peak in female infants with rather constant testosterone concentrations during the first year of life70 might explain why we did not detect any significant associations between 2D:4D and female gender identity. Considering anatomical features of the human brain, the BNST has been linked to gender identity4. An animal study examining mice3 revealed that perinatal metabolites of testosterone masculinize sexually dimorphic cell survival in the BNST, thus suggesting a possible association of perinatal testosterone and gender identity.Nevertheless, our analysis implicates that prenatal androgen exposure plays a role in the development of male gender identity. A possible underlying mechanism may be a discrepancy between the two pre-/perinatal testosterone secretion peaks in males: (1) the first between 10 and 20 weeks of gestation entailing bodily changes and (2) the second during the perinatal window entailing organizational effects in the brain19 reported significant associations of male transgender identity only with R2D:4D but not with L2D:4D, whereas a more recent meta-analysis6 found significant differences between MtF individuals and male controls in both R2D:4D and L2D:4D. Our original investigation, however, revealed significant associations with only L2D:4D but not with M2D:4D or R2D:4D. Our meta-analysis, in turn, points to an association of male transgender identity with (aggregated) 2D:4D in general with subgroup analyses revealing no significant difference between the meta-analytic effect sizes of R2D:4D and L2D:4D. Following these results, we cannot confirm an increased sensitivity to prenatal testosterone exposure in the right hand as often postulated68.A previous meta-analysis8, as explained in the “75. This is—among other things—due to a lack of X-linked inheritance in classical genetic studies77 or in genome-wide association studies (GWAS)78, even though the androgen receptor is located on the X chromosome. Some authors argue that if 2D:4D reflects prenatal testosterone effects and if androgen receptor variants moderate testosterone effects, androgen receptor variants should show systematic relationships with 2D:4D . Additionally, meta-analyses do not confirm an association between 2D:4D and the effectiveness of androgen receptors (genetic CAG and GGC repeat length polymorphism)79. The largest GWAS on 2D:4D so far only shows a slight positive influence of CAG repeats in women, but not in men78. Thus, other mechanisms may also be involved in the development of 2D:4D besides sex hormones and genetics, for example prenatal corticosterone80 and prenatal stress81.In order to draw conclusions about the prenatal androgen load, 2D:4D is used as a proxy47 and by assessing the risk of bias of the studies according to the Newcastle–Ottawa Quality Assessment Scale for case–control and cohort studies48. For the sample of MtF individuals versus male controls, heterogeneity was small, indicating low inconsistency of studies’ results. This includes variation mainly due to sampling error, i.e. to the variation in study outcomes between studies. Our sensitivity analysis suggests that the effect does not depend on any one single study and therefore, is reasonably robust.This analysis provides new case–control data from a comparably large sample size and a meta-analysis including a wide array of dimensions of transgender identity, which we incorporated in separate subgroup analyses. We ensure a standardized procedure for our meta-analysis by following the PRISMA statement entirely19. This additionally entails that our sample is not fully suited for multiple testing and we, consequently, could not control for alpha inflation; hence impeding the explanatory power of our results. Third, in the meta-analysis of FtM individuals versus female controls, heterogeneity was high, indicating systematic variation not caused by the study-related sampling errors and possibly resulting in overestimated effect sizes for this sample. Fourth, there were some methodological difficulties, such as information and selection bias potentially inflating the case–control difference and impeding the generalizability of results, when applying the study design of case–control studies. This extends to the control self-selection via flyers in the original investigation, an approach in which motivational factors related to personality traits or lifestyle are likely to play a role82. Consequently, the control group does not reflect a population-based sample.There are some limitations to this study. First, criticism of the concept of 2D:4D as an appropriate proxy for prenatal androgen exposure has emerged as described above in the “Following our meta-analysis, no recommendation can be made as to whether future 2D:4D examinations should be performed analyzing right or left hand 2D:4D or using methods with or without soft tissue deformation . We recommend conducting original investigations in this field following a case–control design with reliable diagnoses (preferably made by clinicians) and computing a priori power analyses for determining an adequate sample size.In summary, our results strengthen the assumption of a multifactorial etiology of gender identity and suggest that prenatal androgen levels are involved. Nevertheless, deeper analyses of other biological and non-biological mechanisms are required. We therefore recommend further studies on the developmental pathways of 2D:4D and transgender identity.Supplementary Information.


Systemic immune-inflammation index (SII) has been suggested to be effective to reflect the inflammatory status and thus may be an underlying biomarker for prognosis prediction. This hypothesis has been demonstrated in meta-analyses on several cancer types. However, there was no study to confirm the prognostic roles of SII for gynecological and breast cancers, which was the goal of our study.PubMed, EMBASE, and Cochrane Library databases were searched to collect the articles exploring the associations of SII with prognostic outcomes  in gynecological and breast cancers. The prognostic value of SII was estimated by hazard ratio (HR) or relative risk (RR) with 95% confidence interval (CI).P < 0.001), DFS/PFS  and an increased risk for LNM  in patients with gynecological and breast cancers. Subgroup analysis confirmed the prognostic role of SII for OS was applicable to all cancer types, but the association with DFS/PFS and LNM was only significant for ovarian cancer and breast cancer, especially triple-negative breast cancer. No significant association was detected between SII and LVI.Nine articles involving 2724 patients in 11 datasets were included. Meta-analysis showed that a high SII index was significantly associated with poor OS (HR = 2.12, 95% CI, 1.61–2.79, High SII may be a promising indicator for the prediction of poor prognosis in patients with gynecological and breast cancers, especially ovarian cancer and triple-negative breast cancer. Gynecological and breast cancers are the two leading causes of death among women . AccordiP < 0.001; PFS: HR = 7.61, 95%CI = 3.34–17.35; P < 0.001], but also in the discovery cohort  [P < 0.001 [P < 0.00 1[P = 0.045 [P = 0.005 [P = 0.005) and validation  cohorts, the study of Huang et al. supported that SII was an independent risk factor for prediction of OS in cervical cancer patients [In recent years, increasing evidence has shown that activation of inflammation is a crucial mechanism for the recurrence and metastasis of gynecological , 4 and b= 0.001) . Multiva < 0.001 ; Wang et < 0.00 1) and DFS = 0.045 ; Wang et = 0.005 ) in patipatients . Furtherpatients , indicatpatients , 16. TheThis meta-analysis was performed based on the Preferred Reporting Items for Systematic Review and Meta-Analysis (PRISMA) statement. Because the present study was a meta-analysis of articles published previously, ethical approval and patient consent were not required.Two authors independently searched eligible articles in PubMed, EMBASE, and Cochrane Library from the date of establishment to January 1, 2020. The search strategy included  AND (“cancer” or “carcinoma” or “tumor”) AND (“systemic immune-inflammation index” or “SII”). Additionally, references of included publications and reviews were manually reviewed for potential trials.The study selection was completed by two independent investigators. Publications were eligible if they met the following inclusion criteria: (1) the enrolled patients suffered from gynecological and breast cancers which were pathologically diagnosed; (2) patients did not have an active infection, inflammatory, or comorbid diseases or undergo anti-inflammatory medication before blood examination; (3) neutrophil, platelet, and lymphocyte counts were measured prior to any treatments and SII was calculated; (4) the associations between SII and prognostic outcomes of patients were assessed; (5) the HRs with their 95%CIs were reported or could be calculated from raw data; (6) the cut-off value of SII was provided; and (7) articles were published in English. The exclusion criteria were as follows: (1) studies were duplicated or data were overlapped; (2) letters, case reports, editorials, or reviews; (3) non-human studies; and (4) insufficient data for estimating HRs and 95%CIs for prognosis outcomes.Two authors extracted the following data independently: the name of the first author, year of publication, country, sample size, cancer type, study design, treatment, follow-up period, SII cut-off, source of HRs, outcomes, and HR with 95%CI for each outcome. HR based on multivariate analysis was preferentially extracted if available. The Newcastle-Ottawa Scale (NOS) criteria  was usedz test (P < 0.05) with 95%CI . Cochrane’s Q and I2 statistic tests were used to measure the heterogeneity of included studies. P < 0.10 and I2 > 50% indicated the presence of heterogeneity among studies, so that the pooled HR was calculated by a random-effects model; otherwise, a fixed-effects model was applied. Subgroup analyses were also performed by country, sample size, cut-off, cancer type, follow-up length, and source of HR. The values for dividing the subgroup of sample size, cut-off, and follow-up were selected according to the integer value of the median. Publication bias was estimated by Egger’s linear regression test (P < 0.05 indicated a significant publication bias) [The HR and 95%CI of each study were calculated using STATA 13.0  to assess the associations of SII with OS, PFS, and DFS. The relative risk (RR) and 95%CI were calculated for lymph node metastasis (LNM) and lymphovascular invasion (LVI). A pooled HR or RR > 1 indicated a poor prognosis for patients with high SII. Statistical difference was determined by using a on bias) . Publicaon bias) . The robn = 84) and screening titles and abstracts (n = 32), 10 studies were assessed by full text for eligibility. One study was excluded due to lack of relevant data. No additional records were identified through manual searching. Finally, 9 published articles involving 2724 patients were included in this meta-analysis [n = 6), Italy (n = 1), USA (n = 1), or Spain (n = 1). Most of the studies  extracted the HR and 95%CI from the multivariate analysis and only 2 from the univariate analysis [As shown in Fig. analysis , 20–22. analysis , 14 contanalysis , 21, 22 analysis , 15 inveanalysis ; while aanalysis , 21, 22 analysis , 20. TheI2 = 72.0%, P < 0.001). The pooled results indicated that a high SII index was significantly associated with shorter OS in patients with gynecological and breast cancers   Fig. , with HRI2 = 81.3%, P < 0.001) (Table P < 0.001) , sample size , cut-off , cancer type , and follow-up length  ; therefore, a fixed-effects model was used (Table P < 0.001) . Furthermore, in the subgroup with cut-off less than 600, the risk differences of LNM were not statistically significant between high and low SII index (P = 0.094)  showed that SII index could not predict the LVI for patients with gynecological and breast cancers  . For DFS/PFS, publication bias seemed to be present (P = 0.007); however, the prognostic significance of SII remained unchanged  after trim-and-fill adjustment. The sensitivity analyses also demonstrated that the pooled results could not be affected after the removal of any one study  varied by ethnicity and BMI was positively correlated with SII . Hereby,Our findings provide evidence that high SII may be a promising indicator for prediction of poor prognosis  in patients with gynecological and breast cancers, especially ovarian cancer and triple-negative breast cancer.


Previous studies have shown that visit-to-visit blood pressure variability (BPV) is associated with chronic kidney disease (CKD). However, the results have not been consistent among studies. This systematic review and meta-analysis was conducted to comprehensively assess the association between visit-to-visit BPV and the risk of CKD.Medline, Embase, and the Cochrane Library were searched from the date of inception through 1 August 2019 using the terms “blood pressure variability,” “chronic kidney disease,” “nephropathy,” and other comparable terms. The primary outcome was the development of CKD. Two reviewers extracted the data independently. Meta-analysis was performed using a random effects model.Fourteen studies were included in the systematic review and meta-analysis. The risk of CKD was significantly greater in patients with high baseline systolic blood pressure variability (SBPV) than in patients with low baseline SBPV: the standard deviation (SD) showed relative risk (RR) of 1.69 and 95% CI of 1.38–2.08, the coefficient of variation (CV) showed RR of 1.23 and 95% CI of 1.12–1.36, and variance independent of mean (VIM) showed RR of 1.40 and 95% CI of 1.15–1.71. RRs for each unit increase in visit-to-visit SBPV and risk of CKD were 1.05 (95% CI: 1.03–1.07) for SD, 1.06 (95% CI: 1.03–1.09) for CV, and 1.1 (95% CI: 0.96–1.25) for VIM. Diastolic BPV was similarly predictive of CKD based on SD and CV.Increased visit-to-visit BPV might be an independent risk factor for CKD. However, significant heterogeneity was present; thus, future prospective studies are needed to confirm our findings. Our results indicate that treatment of hypertension should control blood pressure levels and prevent abnormal fluctuations in blood pressure to reduce the risk of CKD. Chronic kidney disease (CKD) is a global public health problem that leads to poor health and high costs . A crossHigh blood pressure (BP) has been identified as the leading risk factor for mortality worldwide and one of the most important risk factors for CKD , 4. BP fIt is important to reduce the incidence of CKD by early identification of high-risk patients and modification of treatment-related risk factors. The impact of increases in BPV  on the risk of CKD remains unclear. An improved understanding of this relationship might provide a potential novel target for prevention of CKD. Thus, this systematic review and meta-analysis of cohort studies was conducted to comprehensively assess the association between visit-to-visit BPV and the risk of CKD.https://www.crd.york.ac.uk/PROSPERO; Registration number: CRD42020149248).This systematic review and meta-analysis was conducted in accordance with the Meta-analysis of Observational Studies in Epidemiology (MOOSE) guidelines . ComprehStudies were included if the following criteria were fulfilled: (1) cohort design, including prospective cohort or retrospective cohort, or follow up study of randomized controlled trial; (2) the exposure of interest was BPV; (3) the outcome was CKD ; (4) quantitative estimates were reported for the adjusted relative risk (RR) and 95% confidence interval (95% CI) for CKD associated with BPV.A study was excluded if the following criteria were met: (1) inclusion or exclusion criteria of the study were unclear or unreasonable; (2) insufficient data of interest were reported; (3) the study was a review, supplement, abstract only, commentary article, editorial, or grey literature. If relevant data from ≥2 articles were derived from the same cohort, only the study with a longer follow-up duration or larger population was included in this analysis. Discrepancies were resolved by discussion.Data from all included studies were extracted by two independent review authors (H.L. and J.X.). Discrepancies in data abstraction were resolved through discussion or by a third investigator (W.C.). Data extracted from studies included the following information: first author’s name, publication year, country, database, study design, duration of follow-up, sample size, mean patient age, sex, outcomes, outcome definitions, BPV measurements , fully adjusted RRs, and 95% CIs.The quality in prognosis studies (QUIPS) tool is recommended for quality assessment in reviews of prognostic factors . Two revAll statistical analyses were conducted using Stata version 12 . Risk estimates for each study were reported as odds ratio, risk ratio, or hazard ratio; these were presumed to be approximately equivalent for effect sizes <2.5 and follow-up <20 years and therefore were merged directly , 21, 22.To quantify the dose–response relationship between BPV and risk of renal disease, we calculated the RR for each unit increment in BPV in each study. For studies that reported risk estimates using ranges of BPV, we calculated the midpoint in each category by calculating the average of the lower and upper bounds. When the highest or lowest category was open-ended, we assumed that the open-ended interval was similar in length to the adjacent interval . The gen2 statistic [2 >50% was considered indicative of a significant level of heterogeneity; stratified syntheses and sensitivity analyses were used to explore sources of heterogeneity. Stratified analyses were performed based on follow-up duration, country, population, or sample size. Publication bias was assessed using Egger’s test [Heterogeneity among the included studies was evaluated using the Cochrane Q test and Itatistic . A valuer’s test . A two-tThe results of the literature search are summarized in The characteristics of all selected studies are listed in 2 = 98.8%; S1 Fig in 2 = 91.1%; S2 Fig in 2 = 95.9%; S3 Fig in Visit-to-visit systolic BPV (SBPV) was defined using SD, CV, and VIM. The risk of CKD was significantly greater in patients with high baseline SBPV than in patients with low baseline SBPV  (2 = 71.4%) (2 = 88.4%)  for CKD based on increases in BPV. For each 1 mmHg increase in systolic BP SD, the combined RR for CKD was 1.05 (95% CI: 1.03–1.07) with significant heterogeneity (I= 86.3%) . For eac= 71.4%) . For eac= 88.4%) .2 = 88.7%; S4 Fig in 2 = 95.6%; S5 Fig in 2 = 83.2%) (2 = 71.4%) (Visit-to-visit diastolic BPV (DBPV) was defined using SD and CV. The risk of CKD was significantly greater in patients with high baseline DBPV than in patients with low baseline DBPV  . For eac= 71.4%) .We performed subgroup analysis of the studies based on follow-up duration, country, population, and number of subjects . In thesThere was no evidence of publication bias regarding the BPV SD for CKD, as determined by Egger’s linear regression test . The resTo the best of our knowledge, this is the first systematic review and meta-analysis to explore the potential relationship between visit-to-visit BPV and CKD. Our study demonstrated that high visit-to-visit BPV may indicate an increased risk of CKD. Patients with higher SBPV had a 5% higher risk of CKD for each 1 mmHg increase in SD compared to patients with lower SBPV. We also used CV and VIM to examine the association between SBPV and risk of CKD; the corresponding risk ratios also showed increased risk of CKD. This association remained significant in analysis of the relationship between visit-to-visit DBPV and the risk of CKD. Our results suggest that visit-to-visit BPV is an important clinical factor for predicting CKD.BP is a physiological parameter that reflects hemodynamic status, and it is characterized by significant changes over time. The magnitude of BP variation was initially suspected to interfere with the accuracy of BP status assessments in individual patients. However, BPV is increasingly recognized as an important pathophysiological phenomenon. Visit-to-visit BPV is presumed to reflect a variety of mechanisms, such as fluctuations in activation of the renin–angiotensin–aldosterone system, overactivity of the central sympathetic nervous system, increased secretion of vasoactive compound, and/or increased arterial stiffness, as well as environmental and psychological conditions  . BPV is Even mild renal impairment constitutes a prominent risk factor for cardiovascular disease, infection, cognitive impairment, and reduced physical function . Our finSeveral limitations might have contributed to the significant heterogeneity detected in this meta-analysis. Differences in population, sample size, follow-up duration, incomplete matching, country of origin, or methodology may have caused heterogeneity; of these factors, the observational study design of the included studies might have been the most prominent source. Methodological factors including number of visits, time interval between visits, and assessments by different doctors may have also contributed to the heterogeneity.Taken together, the findings in this systematic review and meta-analysis suggest that increased visit-to-visit SBPV is an independent risk factor for CKD. However, significant heterogeneity was present; thus, future prospective studies are needed to confirm our findings. Our current findings indicate that treatment of hypertension should control BP levels and prevent abnormal fluctuations in BP to restore normal BP rhythm. Additional clinical studies are needed to evaluate the normal reference values of visit-to-visit BPV for clinical practice and identify more effective treatment approaches for reduction of BPV.S1 Data(DOCX)Click here for additional data file.S1 Checklist(DOC)Click here for additional data file.


VDR) gene and the risk of type 1 diabetes mellitus (T1DM) has been evaluated in several studies. However, the findings were inconclusive. Thus, we conducted a meta-analysis to comprehensively evaluate the effect of VDR gene polymorphisms on the risk of T1DM.The association between the polymorphisms in the vitamin D receptor  and 95% confidence intervals (CIs). The methodological quality of each study was assessed according to the Newcastle–Ottawa Scale. To find the potential sources of heterogeneity, meta-regression and subgroup analysis were also performed.All relevant studies reporting the association between VDR gene polymorphisms and T1DM risk. However, the pooled results of subgroup analysis revealed significant negative and positive associations between FokI and BsmI polymorphisms and T1DM in Africans and Americans, respectively.A total of 39 case–control studies were included in this meta-analysis. The results of overall population rejected any significant association between VDR gene polymorphism and T1DM susceptibility in ethnic-specific analysis.This meta-analysis suggested a significant association between  Type 1 diabetes mellitus (T1DM) is a globally-widespread disease that is characterized by a reduction in insulin production or the production of ineffective insulin . T1DM coVitamin D (VitD) is a steroid molecule that has many roles in the body, such as regulation of the immune cells. In addition to immune responses, VitD is also involved in the etiopathogenesis of several disorders, such as cancer, autoimmune disorders, cardiovascular disorders, asthma, and diabetes –9. In anVDR gene which lead to silent mutation associated with increased VDR mRNA stability. In contrast, FokI SNP is located in the start codon that produces a protein with shorter size (424 amino acids), which is more active than the long form (427 amino acids)  and increased risk of T1DM susceptibility in African population under all genotype models; dominant model , recessive model , allelic model , ff vs. FF model , and Ff vs. FF model . Besides, susceptibility to T1DM in Asians compared to Africans and Europeans were not affected by FokI polymorphism , recessive model , allelic model , bb vs. BB model , except Bb vs. BB model  , TaqI (rs731236), BsmI (rs1544410), and ApaI (rs7975232) and susceptibility to T1DM. The findings of meta-analysis on 39 case–control studies, containing 29 studies with 3723 cases and 5578 controls for FokI, 20 studies with 1837 cases and 1895 controls for TaqI, 34 studies with 4826 cases and 7159 controls for BsmI, and 24 studies with 2436 cases and 4074 controls for ApaI, indicated no significant association of VDR gene polymorphisms with T1DM risk in overall population. That notwithstanding, the subgroup analysis resulted in identification of significant associations between FokI and BsmI polymorphism and T1DM in African and American population. Our study provided some beneficial points over previous studies. First, this meta-analysis included further studies with more sample size compared with the previous studies, conferring more conclusive results. Second, we performed subgroup analysis by ethnicity to indicated association of VDR gene polymorphisms with T1DM risk in different ethnical groups.In this study, we performed a systematic review and meta-analysis to achieve a vivid and exact approximation of the associations between the VDR gene polymorphisms and risk of T1DM throughout various populations, resulting in conflicting findings  and susceptibility to T1DM. The findings of meta-analysis revealed no significant association of VDR gene SNPs with T1DM risk in the overall population. However, the subgroup analysis indicated significant associations between FokI and BsmI polymorphism and T1DM risk in African and American population. As a limitation, we did not evaluate a number of VDR gene SNPs that might act in interaction with environmental factors to determine the fate of T1DM pathogenicity. Further investigations on the VDR, above and beyond the genetic as well as traditional risk factors, may confer a possibility for identification of critical susceptibility factors in the disease development, which might be applicable in the personalized medicine for better and optimized therapy of T1DM patients.In conclusion, this study was a systematic review and meta-analysis of 40 case–control association studies to come up with the clear estimation of the associations between the


I2 = 55.4%;p-value = 0.008). The subgroup analysis revealed that only GG has significant association with MetS  without significant heterogeneity . In conclusion, this meta-analysis provides strong evidence that only some SNPs of CRP gene are associated with the risk for development of MetS; and this relationship does not exist in different ethnic populations.Polymorphisms in the C-reactive protein (CRP) genes might have crucial role in the development of metabolic syndrome (MetS). In the current comprehensive meta-analyses, we aim to provide a quantitative assessment of the association between CRP single-nucleotide polymorphisms (SNPs) and the risk of MetS. An electronic search was performed on several databases. After data extraction, random effect model was used to calculate the pooled odds ratio (OR) and 95% confidence intervals (CIs). Four independent studies including case–control, cohort, and cross-sectional methods were analyzed. Our meta-analysis indicated that CRP polymorphisms are not significantly associated with MetS  with significant heterogeneity ( Metabolic syndrome (MetS) is a growing public health problem worldwide. It is defined as a complex syndrome with coexistence of multiple cardiovascular risk factors consisting of atherogenic dyslipidemia, central obesity, proinflammatory and prothrombotic states, hypertension, and hyperglycemia.Ridker et al reported slight increase in moderate C-reactive protein (CRP) levels in relation to the number of MetS components in a follow-up study on healthy American women.Many recent studies evaluated the genetic association between CRP single-nucleotide polymorphisms (SNPs) and MetS in different populations.The electronic databases, such as Science Direct, PubMed, Web of Knowledge, Scopus, and Springer, were systematically searched for all available articles, without any limitation for publication date, language, article type, access type, and species until January of 2019. Articles with the following search terms in the titles, abstracts, or keywords were included: (((“C-reactive protein” OR CRP)) AND (metabolic syndrome OR MetS)) AND polymorphism. We also improved the search by revising the reference lists of all of the selected publications and finding supplementary documents related to these articles.For identifying suitable articles, the following criteria were included in this meta-analysis study. Any article published as an original study that considered the association between CRP polymorphisms and MetS. The available data for each polymorphism was reported and also necessary data to estimate the odds ratio (OR) and 95% confidence interval (CI) was provided. Furthermore, the reviews, animal studies, comments or editorials, overlapped articles, or studies with insufficient information on genotypes or allele frequencies were excluded.The original information of search articles were extracted by two authors (M.M. and M.S.) independently using a reliable and standardized method. The following information was collected from each study: first author, year of publication, origin of the subjects, study design, variables adjusted for the analysis, and the Hardy–Weinberg equilibrium value of the overall association for each SNP of CRP gene.Ztest was used for determining the significance of the pooled OR. Statistical analyses were performed by STATA software . Ap-value of < 0.05 was considered as statistically significant. The heterogeneity among studies was detected using the chi-square test-basedQstatistic, and then it was quantified using theI2statistic.The association between polymorphisms of CRP gene and MetS was evaluated. Statistical analysis was performed by pooled ORs and associated 95% CIs. Thep > 0.05 andI2 < 50%). When heterogeneity was significant, random-effects model (the DerSimonian–Laird method) was used. Finally, for evaluating the existence of heterogeneity among groups, sensitivity analyses were conducted. In addition to calculating publication bias, the Begg's and Egger's tests were applied, in whichp-values of less than 0.05 indicate significant publication bias.The Mantel–Haenszel fixed-effects model was considered when heterogeneity was not significant . We used funnel plot and Egger's test to assess the publication bias. Based on both funnel plot (p = 0.44).The forest plot for association between CRP gene polymorphism with MetS according to all the studies is shown inel plot  and Eggp-value = 0.01) with no significant heterogeneity   . It is Various investigations have reported the association of CRP polymorphisms and MetS; however, the findings of previous studies have been controversial. To start, we chose the CRP gene because there is strong evidence that it is a powerful predictor of incident MetS events. As a result, the most common variants studied were the following polymorphisms: rs1800947, rs3093068, rs3091244, and rs1205 of the CRP gene. In the current research work, to prove whether CRP polymorphism is responsible for the development of MetS, we conducted a meta-analysis, which included four independent studies. Based on literature search, this article is the first comprehensive meta-analysis to evaluate the possible correlations of CRP polymorphisms in MetS.In adult populations, the association between the CRP rs3091244 and MetS has been evaluated in several studies.Some studies reported the association between different populations. For example, in 2010, Hsu et al demonstrated the correlation of rs3091244 on the risk of MetS in Taiwanese adults. This study showed that the non-CC genotypes are associated significantly with MetS after adjustment for smoking, age, sex, and body mass index.Some limitations of our meta-analysis study should be noted. First, the limited number of articles for most of the CRP polymorphisms was available. Second, the languages of the publications were limited to English. Third, our meta-analysis was based on evaluations without adjusting the data for population characteristics such as ethnicity, age, etc. Fourth, due to lack of access to detailed data used in studies, our meta-analyses are based on the various data of CRP SNPs, and finally, it must be noted that the study design of included articles in this meta-analysis was different .Despite these limitations, the current study also has some advantages. First, the statistical power of the analyses was considerably increased as a large number of samples were derived from different studies. Second, in our analyses, we investigated different ethnicity population.In summary, this meta-analysis demonstrated considerable evidence that there were no correlations between various polymorphisms of CRP gene  and the risk of MetS in various populations. Moreover, it is recommended that future researches into the mutual effects of the environment and genes may improve existing concept of the associations between polymorphisms of CRP gene and developing MetS. Further work will help explain the medical and biotechnological implications of these associations.


Background: Polycystic ovarian syndrome (PCOS) is a kind of common gynecological endocrine disorder. And the mutations of melatonin receptor (MTNR) genes are related to the occurrence of PCOS. But previous researches have shown opposite results. So, the object of our systematic review and meta-analysis is to investigate the relationship between MTNR 1A/B polymorphisms and PCOS.Methods: PubMed, Embase, Ovid, the Cochrane Library, Web of Science and three Chinese databases  were used to retrieve eligible articles published between January 1980 and February 2020. And we used the odds ratio (OR) and its 95% confidence interval (CI) to investigate the strength of the association by six genetic models, allelic, codominant (homozygous and heterozygous), dominant, recessive and superdominant models. Review Manager 5.3, IBM SPSS statistics 25 and Stata MP 16.0 software were used to do this meta-analysis.Results: Our meta-analysis involved 2553 PCOS patients and 3152 controls, for two single nucleotide polymorphisms (rs10830963 C> G in MTNR1B and rs2119882 T> C in MTNR1A) and significant associations were found in some genetic models of these single nucleotide polymorphisms (SNPs). For rs10830963, strongly significant was found in the heterozygote model . Additionally, a slight trend was detected in the allelic (G vs. C), homozygote (GG vs. CC) and dominant (GG+GC vs. CC) model of rs10830963 (P=0.05). And after further sensitivity analysis, a study with high heterogeneity was removed. In the allelic (P=0.000), homozygote (P=0.001), dominant (P=0.000) and recessive  model, strong associations between rs10830963 and PCOS were found. Moreover, for rs2119882, five genetic models, allelic , codominant  and heterozygote model , dominant  and recessive model  showed significant statistical associations with PCOS.Conclusion: MTNR1B rs10830963 and MTNR1B rs2119882 polymorphisms are associated with PCOS risk. However, the above conclusions still require being confirmed by much larger multi-ethnic studies. Moreoveon (IVF) , slow thon (IVF) , increason (IVF) .FTO) gene [FSHR) gene [VDR) gene [Additionally, multiple studies reveal that melatonin secretion is increased in polycystic ovarian syndrome (PCOS) patients . PCOS isTO) gene ,30, FollHR) gene ,32, DENNHR) gene , VitaminDR) gene  and so oAdditionally, the SNPs of MTNR 1A/B gene (MTNR1A/B) are found to have associations with many kinds of metabolic disorders. For example, MTNR1B rs10830963 polymorphism is not only associated with fasting blood glucose levels, but also with IR and T2DM risk . And MTNOur study was conducted with the requirements of Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) . All datFive English databases  and three Chinese databases  were searched in our study. And we used three subject heading terms, ‘Polycystic ovarian syndrome’, ‘melatonin’ and ‘melatonin receptor’, as well as the synonyms of these terms confirmed by Medical Subject Headings (MeSH) referred to in the Supplementary Material S1 (Search strategies) , where oAs the definition of PCOS used in the present study, all the eligible studies adopted the 2003 Rotterdam diagnostic inclusion criteria of PCOS : (1) thiAssess the association between SNP in MTNR and PCOS risk,Contain particular data, such as the frequency of MTNR genotypes.Included references needed to meet the following criteria:Abstracts and reviews,Repeated dissertation,No research on genotype distribution,Non-human research.Excluded studies should meet the following criteria:http://www.ohri.ca/programs/clinical_epidemiology/oxford.htm) [Among them, Newcastle–Ottawa Scale (NOS), available from URL (ord.htm) , was perAll studies retrieved from all the databases were assessed and reviewed by the first and second authors (Shiqi Yi and Hao Shi) based on eligibility criteria. The data extraction results were agreed upon discussion with another first author (Jiawei Xu). Study and descriptors such as author, publication time, country, ethnicity, study period, study type, NOS scores, the number of participants and methods of study were extracted and recorded from the included studies.The third and fourth authors (Wenbo Li and Qian Li) independently assessed the quality of eligible studies using NOS which consisted of three parts, namely selection, comparability and exposure, and their corresponding eight scoring items. Except for comparability which could score up to two stars, each item could only score up to one star. Possible maximum number of stars in each study was 9. When the score of eligible study in NOS was ≥7, the present study was regarded as high-quality research . When schttps://community.cochrane.org/help/tools-and-software/revman-5) was performed for data analysis [2 test) by IBM SPSS statistics 25 software [2I, a test statistic, for quantification [P≥0.1 or 2I ≤ 50%, the fixed-effects model could be established to analyze. Otherwise, the random-effects model could be used. Next, a Z-test was performed to evaluate the statistical significance of combined ORs in quantitative synthesis. And all analysis will use 0.05 as significantly statistical level. Finally, by eliminating the heterogeneous merger studies one by one, the effect of a single heterogeneous study on the entire model was further discussed through a sensitivity analysis. And Stata MP 16.0 software was used to analyze the publication bias by Egger’s and Begg’s tests.Review Manager 5.3 software  and their recommended links. Then, all searched reference lists were imported into EndNote X9, which helps to remove duplicate studies. The majority of articles were excluded for unrelated titles and abstracts and other eight articles were excluded because of which studied populations were not PCOS patients, genotype data were not valid, or references were unrelated. Additionally, Song et al. (2015) study  was remoThe eligible references in the present study involved two genes and six polymorphisms, MTNR1B gene  and MTNR1A (rs2119882). Four references, including Xu et al. (2019) , Yang etThe results of the association between the MTNR polymorphisms and the PCOS risk are shown in P=0.02, 2I = 44%). In the allelic, homozygote and dominant model, the variations were weakly associated to PCOS risk . In the recessive model, this mutation had a weaker association . And in last model, the superdominant model, no statistically significant association were found . All studies were evenly distributed on both sides of the center line, and no obvious publication bias was found.To study the relationship between rs10830963 (C>G) and PCOS risk, we included four case–control studies  and analP<0.001, 2I = 54%; OR: 2.07, 95% CI: 1.55–2.76, P<0.001, 2I = 34%; OR: 1.32, 95% CI: 1.05–1.65, P=0.02, 2I = 56%; OR: 1.59, 95% CI: 1.05–2.41, P=0.03, 2I = 69%; OR: 1.72, 95% CI: 1.34–2.22, P<0.001, 2I = 0%). However, in the superdominant model, no significant statistical association was found . All models had low or moderate heterogeneity, so the fixed-effect models were used to analyze them.To study the association between rs2119882 and PCOS risk, we included two case–control studies  and used2I and P≤0.01), a sensitivity analysis was conducted by removing all the eligible references one by one in each model to assess the influence of each reference heterogeneity on the whole model analysis [P=0.000, 2I = 70%; OR: 2.78, 95% CI: 1.51–5.10, P=0.001, 2I = 73%; OR: 1.58, 95% CI: 1.31–1.90, P=0.000, 2I = 50%; OR: 2.30, 95% CI: 1.40–3.77, P=0.001, 2I = 67%).For genetic models with high heterogeneity  conclusions are more heterogeneous [Melatonin works by activating two important G protein-coupled receptors, MT1 and MT2 . By querogeneous , multiplogeneous . Therefoogeneous ,63 and togeneous .In our meta-analysis, we find that in the heterozygote model, MTNR1B rs10830963 polymorphism is significantly associated with the PCOS occurrence. Through the sensitivity analysis, after excluding Wang et al. study , multiplMoreover, obese women in PCOS patients can be complicated with T2DM . At the As a result of weak endocrine regulation of melatonin through mutation receptor, to some extent, our meta-analysis supports the feasibility and acceptability of appropriate melatonin supplementation for PCOS ,69. At tThe SNPs rs10830963 and rs2119882 of MTNR1B and MTNR1A, are associated with a much higher PCOS risk in Chinese population. And above conclusions still require confirmation by future study including much larger multi-ethnic studies and other SNPs instead of the above conclusions.


Cancers are the first main cause of premature death in developed countries. Since brain tumors, especially gliomas, are the most lethal type of cancers, risk factors for their prevalence are still being discussed. Nearly 30–50% of all cancers could be prevented by proper nutritional habits and other lifestyle factors, but their influence on the tumors of the central nervous system has not been explained completely and still requires further studies. That is why we attempted to review the available research in this field, with a special focus on the factors with the proven protective activity observed in other cancers. Adequate vegetables and antioxidants (such as vitamins C and A) provided with a diet could have a protective effect, while other factors have shown no correlation with the incidence of glioma. However, further studies are necessary to determine whether fish, coffee, and tea consumption may prevent glioma. Maintaining proper body weight and undertaking a sufficient level of daily physical activity also seem to be important. Excessive body mass index (BMI) and higher attained height have increased the risk of glioma. In order to link more accurately the chosen factors to the prevalence of gliomas, it seems necessary to conduct large cohort, prospective, controlled studies in different world regions. Cancers are the first main cause of premature death in developed countries; this has been confirmed in 48 (28%) of 172 study countries among people under the age of 70, based on the epidemiological data of the World Health Organization (WHO). According to global cancer statistics assumptions in 2018, there were reported 18.1 million new cancer diagnoses and 9.6 million cancer deaths. Lung cancer and breast cancer were diagnosed most frequently—each one constituted 11.6% of new cases. The highest mortality of all cancer deaths was caused by lung cancer (18.4%). Regarding the available statistics for both sexes, brain tumors are not included in the 10 most common cancers and in the 10 cancers that cause the highest number of deaths. Despite this fact, the comparison of the age-standardized rate data for tumors incidence of the brain and the central nervous system (CNS)  and mortality  indicates extremely poor prognosis for this group of patients [However, in the group of brain tumors, in particular, malignant gliomas are the most lethal type of cancer. In 2019, a total of 296,851 new cases of cancers of the brain and the central nervous system were diagnosed, of which 52.5% were confirmed in Asia, 21.8% in Europe, 10% in Latin America and Caribbean, 9.1% in North America, 5.7% in Africa, and 0.82% in Oceania. The cumulative risk (CR) for the incidence of these types of cancers is the highest for Southern Europe and has been calculated as 0.65% and is the lowest for Western Africa, i.e., 0.11%. The CR in Australia and New Zealand has been established to be 0.56%, in North America as 0.52%, and 0.48% in South America, whereas, in Asia regions, the greatest CR has been determined in the Western Asia—0.49%. The CR of very high human development index countries (HDI) is higher than the CR of low HDI countries  [glioblastoma multiforme) and are considered secondary GBMs. Primary GBMs develop de novo and are molecularly distinct from secondary GBM. GBMs are the most common astrocytic glioma and are highly aggressive and lethal [Gliomas are often diagnosed in the advanced stage and are poorly responsive to currently available treatment. Glioma is the most prevalent type of CNS tumor in adults and constitutes 70% of all brain tumors . Based sd lethal . Fast-grd lethal . The medd lethal .The current treatment regimen varies depending on the tissue diagnosis and the selected molecular markers assessment. Younger age and better performance status are considered as therapy-independent prognostic factors. Whenever feasible, neurosurgical removal of the tumor with a safety margin for healthy tissue to improve neurological function is recommended in all newly diagnosed patients. The schedule of radiotherapy is also determined with regard to diagnosis and aims to improve the local control and preserve the patient’s function. Additionally, chemotherapy is included in the standard care for most glioma patients. The most commonly used cytotoxic agents are temozolomide, procarbazine, and lomustine. Due to difficulties associated with complete resection of glioma, the actions taken to prevent its progression play an important role in the treatment of this type of brain tumor . To date, 25 risk factor loci and several rare inherited mutations that may cause glioma in some families have been identified, but ionizing radiation is the only confirmed environmental risk factor . It is eThis paper was based on the articles published in English-language journals until December 2019, available in three electronic databases: Scopus, PubMed, and Google Scholar, which meet required quality standards in relation to the information on the dietary and lifestyle factors discussed in this paper and their association with the prevalence of gliomas. The articles were searched by using the following keywords: “glioma”; “glioblastoma”; “brain tumor” in collocations with “diet”; “nutrition”; “lifestyle factors”; “alcohol”; “tea”; “coffee”; “antioxidants”; “vitamin A”; “vitamin C”; “vitamin E”; “fruits”; “vegetables”; “meat”; “red meat”; “n-nitroso compounds”; “fish”; “fatty acids”; “physical activity”; “body weight”. We adopted the following inclusion criteria: the research in the study topic, specifying glioma cases, with the full text available. The individual studies incorporated in the meta-analysis on the selected topic were excluded from the overall review.Alcohol consumption is one of the strongest lifestyle-related carcinogens. It is considered to be a risk factor for brain tumors because of its ability to cross the blood-brain barrier and a neurotoxic effect. This may lead to structural changes in the hippocampus in adolescence and decreased brain volume (white and grey matter) in adulthood, probably contributing to the changes in the axonal integrity and myelination . The prop = 0.001) and overall CNS tumors  and alcohol intake [The negative impact of alcohol consumption has been demonstrated not only in the pathogenesis of selected tumors but also in cancer survivors. In the large meta-analysis study, focusing on the assessment of dietary factors in the patients cured of cancer, drinking alcohol was positively linked to the overall mortality (relative risk (RR): 1.08, 95% CI: 1.02–1.16) . Neverthl intake . The invl intake  on threel intake .Despite the fact that alcohol is an important risk factor, increasing the risk of developing certain types of cancer, based on the results above, alcohol consumption does not seem to enhance glioma risk significantly, but further investigations are required to confirm these conclusions.The mechanisms by which alcohol could induce cancerogenesis is still not clear; nonetheless, the cellular stress, the genotoxic effect of acetaldehyde, inflammation, and altered folate metabolism are considered . HoweverCoffee, a rich source of many health-promoting compounds, is the drink consumed most frequently in the world. However, it also contains dangerous substances, which are produced in the roasting process . Roasted coffee contains more than 1000 biologically active compounds, part of which could be beneficial for humans, characterized by the antioxidant, antifibrotic, antimutagenic, and anticancer activity . CaffeinCamellia sinensis leaves. Three basic types of tea are distinguished , depending on the process carried out. Green tea, unlike black and oolong, is not fermented. Tea is a rich source of polyphenols (including catechins). However, special attention should be given to green tea catechins amid which epigallocatechin-3-gallate (EGCG) is the most common, while epicatechin-3-gallate (ECG), epigallocatechin (EGC), and epicatechin (EC) are found in smaller quantities. Many mechanisms of green tea catechins have been described, which incorporate anti- and pro-oxidant activity, apoptosis induction, transcriptional factors inhibition, influence on the immune system, epigenetic alternation, inhibition of the receptor tyrosine kinase pathways, and anti-metabolic syndrome effects [Tea is another popular drink that is often consumed. Its production involves processing  effects . EGCG ma effects .p = 0.02) and in females (p = 0.05) but not in males (p = 0.22) [In 2013, Malerba et al.  conducte = 0.22) .2+ release channel— inositol 1,4,5-trisphosphate receptor (IP3R) subtype 3—and, consequently, slowed glioblastoma invasion and migration [The individual analysis indicated no unequivocal relationship between the consumption of coffee, tea, or combined drink of both beverages and risk of glioma. According to the data above, none of the studies has found that tea and coffee consumption increases the risk of glioma, but more research is required to prove whether they truly can prevent this type of cancer. As mentioned at the outset, coffee and tea ingredients generally have an anti-cancer effect. However, there are only a few studies regarding the anti-glioma activity. In the laboratory study on the mouse xenograft model, caffeine, which is the central nervous system stimulant, reduced the tumor mass and prolonged survival. In the in vitro model, it was demonstrated that caffeine inhibited the Caigration .Further studies are necessary to prove and explain the mechanisms of the inhibitory influence of bioactive compounds on glioblastoma development.High intake of fruits and vegetables  is one of the elements of a healthy lifestyle suggested to reduce the risk of cancer incidence. In the meta-analysis, taking into consideration the random relations, Aune et al.  observedp < 0.001). In the analysis of individual tumor histology subtypes, the researchers determined that these associations were strongest in astrocytomas and glioblastomas [Ying  conductelastomas .Protection and repair of DNA damage, modulation of DNA methylation, induction of phase II detoxifying enzymes, and promotion of apoptosis are mentioned among the mechanisms responsible for the protective effect of eating fruits and vegetables. The above mechanisms of fruits and vegetables action are attributed to the presence of numerous anti-tumor compounds, such as vitamins, flavonoids, isothiocyanates, glucosinolates, and dithiolthiones .Although the protective effect of eating vegetables has not been demonstrated in all individual studies, the results of the meta-analysis carried out by Ying deserve attention. A sufficiently high intake of vegetables could be an element of glioma prevention, especially among Asians. In turn, the beneficial influence of fruits intake has not been sufficiently documented for this type of cancer. However, it is vital to consume an appropriate amount of fruits because of their preventive role in reducing the incidence of other types of cancers.Eicosapentaenoic acid (EPA) and docosahexaenoic acid (DHA), included in polyunsaturated fatty acids n-3 (PUFAs), have numerous positive effects on the human body function. PUFAs should be provided with a diet because the human body can biosynthesize them only in small amounts. The dietary source of EPA and DHA is seafood .n = 1,262,104, glioma cases: 2313). Fish consumption had no effect on glioma risk, both in overall follow-up and after excluding the first 5 years of follow-up [Due to protective EPA and DHA effects on brain tissues, it can be suggested that fish intake could also reduce glioma incidence. In the meta-analysis, Lian et al.  looked follow-up , the datollow-up .The balance between PUFA n-3 and PUFA n-6 seems to play a crucial role in the formation of pro- and anti-inflammatory lipids. Recently, attention has been paid to pro-cancerogenic effects of omega-3 fatty acids, but this relationship could exist solely in individuals with specific genetic or metabolic predispositions. It is unclear whether the pro-cancerogenic effect is caused by pollution found in the marine food or by the imbalanced proportion of n-3:n-6 fatty acids in a daily diet .Having reviewed the currently available literature, it seems that fish consumption has no significant impact in the case of glioma development, but the protective effect of fish consumption in other brain cancer suggests a possible protective activity against cancerogenesis in the brain tissues. However, further research should be carried out to explain this mechanism.According to IARC, processed meat has been classified as cancerogenic to humans, and this effect refers to colon and rectum cancer. Red meat consumption has been determined as probably cancerogenic to humans, especially regarding colon, rectum, pancreas, and prostate cancer . The canA meta-analysis conducted by Wei et al. , taking Ward et al.  studied In turn, no relations were shown between the NOC precursors, such as nitrate and nitrite intake, and the glioma risk in the meta-analysis by Wang et al. . They inBased on the available literature, it seems that iron intake (both heme and non-heme) has no influence on the glioma incidence risk. However, taking into account the results of meta-analyses, some studies seem to show a relationship between the consumption of red processed meat and the risk of glioma development, while other studies have shown no relationship. The correlation between the red unprocessed meat consumption and the incidence of glioma has not been proved explicitly.Many compounds that could have a protective, delaying, or eliminating effect on oxidative damage in molecules are classified as antioxidants with different action mechanisms. Enzymes, such as catalase or glutathione peroxidase, and metal-binding proteins like ferritin and minerals  are involved in the prevention of free radicals formation. Another group of antioxidative compounds is focused on preventing oxidative chain reactions by capturing free radicals. Such a mechanism of action is observed, among others, in the case of glutathione, vitamin E, vitamin C, and carotenoids, while the damage caused by free radicals is repaired by lipases, proteases, DNA repair enzymes, or similar compounds .n = 2260 RR: 0.85, 95% CI: 0.63–1.16; cohort n = 920 RR: 1.00, 95% CI: 0.77–1.13), no significant effect on the level of vitamin E intake was determined, either [Vitamin E involves a group of compounds belonging to tocopherols  and tocotrienols. Antioxidative mechanisms of vitamin E  properties include the elimination of free radicals, induction of cell apoptosis, and an inhibitory effect on tumor growth and cancerogenesis . Besides, either . The dat, either , which w, either , median p < 0.001). However, no similar dependencies were observed in other countries [Vitamin A includes a group of compounds characterized by unsaturated hydrocarbons, in which retinol and its derivatives are included . Among antioxidative mechanisms of vitamin A activity, preventing lipid peroxidation, regulating cell differentiation, and inhibition of cancer cell proliferation are also described, besides its participation in the induction of detoxifying enzymes. Specific binding proteins for the retinol and retinoic acid could work as hormones controlling cell differentiation and influence the transport of these compounds across the nuclear membrane. Regarding provitamins A, the β-carotene is most effectively converted to retinol, and its main property is to reduce free radicals and prevent their negative effects . In the ountries .The next antioxidant vitamin, largely responsible for neutralizing free radicals, is vitamin C, which is most common in fruits and vegetables. One of the known anticancer mechanisms of vitamin C is the inhibition of nitrosamines formation by blocking the nitrosation reaction. Besides, vitamin C stimulates the immune system, participates in the protection against oxidative DNA damages, and inhibits the activation of carcinogenesis, especially, at the phase of tumor promotion . It was p = 0.05) [Zinc has also been proved to possess antioxidant properties. It takes part in protection against oxidative stress, DNA replication, and protein synthesis and is an important micronutrient for brain health. Zinc is a crucial element for human metabolism, being a part of more than 300 enzymes, which are responsible for several biological processes. Its deficiency could result in enhanced oxidative stress, leading to damage in DNA, proteins, and lipids. It is suggested that inadequate serum levels of zinc could be correlated with cancer incidence, especially in the development and progression of prostate, pancreatic, and breast cancer . In the  = 0.05) .With regard to the studies on the effect of antioxidants on the cancer patients’ survival, the previous studies have shown divergent results since it has been demonstrated that antioxidants can increase the tumor resistance to chemotherapy or radiotherapy, which could contribute to the shorter survival time or antioxidants might reduce the toxicity of treatment in relations to healthy tissues, which on the other hand, would improve the patients’ survival ,53. Brain = 497) compared to the controls in the case of the total antioxidant index (p < 0.05) and daidzein intake (p < 0.05) with a diet. While analyzing the data for all cases (n = 802), it was demonstrated that the total antioxidant index (p < 0.003), carotenoids (p < 0.05), and phytoestrogens—daidzein (p = 0.003), matairesinol (p < 0.05), secoisolariciresinol (p < 0.003), and cumasterol (p < 0.003)—were statistically significantly inversely associated with glioma risk [In turn, Tedeschi-Blok et al.  studied oma risk .p = 0.009). In the study group, tumor progression time was longer , though not statistically relevant. Moreover, after completing the research, Puri et al. [p = 0.05). It has been suggested that lycopene supplementation may be used as an adjunct to glioma treatment, but these findings require confirmation in the larger studies using the current treatment standard.Lycopene has been proven as an anti-invasion and pro-apoptotic compound, which could cross the blood-brain barrier . Nutritii et al.  found thThere is limited but positive evidence of the role of vitamin A and C in decreasing the glioma risk. This protective effect exists, especially among Americans, since most of the studies discussed in the meta-analyses above were carried out in the USA. Therefore, it seems rational to emphasize the role of proper nutrition in the prevention of gliomas, particularly the consumption of food products that provide adequate amounts of these antioxidants.Many dietary factors are related to the selected anthropometric indicators. Anthropometry is commonly used to measure the general nutritional status and predict health, performance, and survival of individuals. The indicators most frequently used are weight, height, and circumferences . Among determinants of adult’s height are genetic and environmental factors, of which the childhood nutritional status seems to play an important role. Higher levels of insulin-like growth factor (IGF) theoretically may be responsible for linking the growth achieved with the occurrence of individual cancers. Importantly, the height cannot be modifiable, in contrast to body weight. The results from the Million Women Study (MWS) showed that tallness could be seen as a potential risk factor for glioma development in adults; the risk of glioblastoma increased by about 20% for every additional 10 cm of growth . Moore e2 or over 30 kg/m2) is one of the causes of several cancers: colon and rectum, endometrium, ovary, esophagus, gallbladder, pancreas, liver, prostate (advanced), thyroid, multiple myeloma, and renal cancer [Excessive body weight is considered a risk factor not only for the development of several types of cancer but also for the survival and prognosis of cancer patients. Statements of two organizations—IARC and WCRF (World Cancer Research Found)—are consistent and indicate that overweight or obesity  was related to a slightly elevated glioma risk , whereas adult BMI and circumference of waist had no significant association with glioma incidence. Furthermore, the researchers determine no pre-diagnostic changes in body weight for 8 years before diagnosis [2) at the age of 21 in the study conducted by Little et al. [2) in women increased the risk of glioma , and a statistically relevant relationship was reported in the group of overweight females , while in the case of obese women, there was no such relationship. In the studies on men and the studies not taking into account the gender division, no significant relationship was found between overweight, obesity, or overweight and obesity combined and the risk of developing glioma [It must be highlighted that obesity was the strongest lifestyle factor in NIH-AARP Diet and Health study, enhancing the glioma risk to the greatest extend. Adults, who were obese at the age of 18, had nearly 4 times increased glioma risk in comparison to participants with normal weight  . Similariagnosis . Excessiiagnosis  meta-anaiagnosis . Howeveriagnosis . The metiagnosis . On the e et al.  was assoe et al. . Sergente et al.  meta-anag glioma .n = 68,132; observational study, n = 93,676), and compared the relationship between the anthropometric indicators and the risk of glioma in postmenopausal women aged 50–79. At the time of entering the study, all participants were subjected to anthropometric measurements carried out by the trained staff, while the data on anthropometric indicators from the years preceding the study were taken solely from the observational study . In Kabat et al.’s study [Kabat et al. , in turn’s study , there w’s study .Hence, based on the results above, the time of occurrence of excessive body weight seems to be important. Being overweight or obese before the age of 18  may increase the risk of developing glioblastoma, while the excessive body weight achieved in adulthood seems to have a little impact.The levels of physical activity for recommended adults (18–64) to maintain health include undertaking at least 150 min of moderate-intensity aerobic activity or 75 min of vigorous-intensity aerobic activity per week. Increasing the time of types of physical activity mentioned above could additionally be beneficial. It is also suggested to include strength training twice a week involving major muscle groups . The impAlthough we have no influence on height attained, it seems important to pay attention to maintaining proper body weight, especially in adolescence, to reduce the glioma risk. For this reason, taking an appropriate level of daily physical activity could also be beneficial in the prevention of glioma development.A summary of the aspects reviewed in this paper is presented in Considering the facts presented above, it is difficult to associate nutritional and lifestyle factors with the incidence of gliomas due to their complex etiology. Dietary factors that may significantly affect the risk of developing glioma are still not clearly identified and described. Based on the available literature, it can be assumed that coffee and tea consumption and the intake of adequate vegetables and antioxidants (such as vitamins C and A) could have a protective effect. While other factors like alcohol intake, red meat, and red processed meat consumption, as well as iron intake (both heme and non-heme), have shown no correlation with the incidence of glioma. In turn, the consumption of fish, coffee, and tea does not increase the risk, but further studies are necessary to determine whether they may prevent glioma. Maintaining a proper weight of the body, as well as undertaking a sufficient level of daily physical activity, also seems to be important, especially during adolescence. Higher BMI has increased the risk of glioma. The attained height, which cannot be modifiable in contrast to body weight, has shown a positive association with glioma development. However, there is still an insufficient number of studies in the field of nutritional factors and the risk of glioma. In order to link more accurately selected nutritional factors to the incidence of gliomas, it seems necessary to conduct large, prospective, controlled studies in the different world regions.