Though a plethora of functional magnetic resonance imaging (fMRI) studies explored the neurobiological underpinnings of borderline personality disorder (BPD), findings across different tasks were divergent. We conducted a systematic review and activation likelihood estimation (ALE) meta-analysis on the fMRI studies conducted in BPD patients compared to healthy controls (HC). We systematically searched PubMed and PsychINFO from inception until July 9th 2020 using combinations of database-specific terms like ‘fMRI’, ‘Neuroimaging’, ‘borderline’. Eligible studies employed task-based fMRI of the brain in participants of any age diagnosed with BPD compared to HC, during any behavioral task and providing a direct contrast between the groups. From 762 entries, we inspected 92 reports full-texts and included 52 studies (describing 54 experiments). Across all experiments, the HC > BPD and BPD > HC meta-analyses did not yield any cluster of significant convergence of differences. Analyses restricted to studies of emotion processing revealed two significant clusters of activation in the bilateral hippocampal/amygdala complex and anterior cingulate for the BPD > HC meta-analysis. Fail-safe N and single study sensitivity analysis suggested significant findings were not robust. For the subgroup of emotional processing experiments, on a restricted number of experiments providing results for each group separately, another meta-analysis method (difference of convergence) showed a significant cluster in the insula/inferior frontal gyrus for the HC > BPD contrast. No consistent pattern of alteration in brain activity for BPD was evidenced suggesting substantial heterogeneity of processes and populations studied. A pattern of amygdala dysfunction emerged across emotion processing tasks, indicating a potential pathophysiological mechanism that could be transdiagnostic. The disorder has a considerable prevalence (5.9% lifetime2) and is associated with significant and widespread impairment of patients’ lives5. A plethora of neuroimaging studies6, most of which employed functional Magnetic Resonance Imaging (fMRI), attempted to delineate the neurobiological underpinnings of BPD. However, findings across different types of tasks were divergent. For example, some studies showed activation increased in the amygdala8, insula9, occipital, frontal and temporal areas11, while others reported decreased activation in both frontal and temporal regions12, cingulate cortex and nucleus accumbens13. Yet other studies found no significant differences in activation between BPD and healthy subjects (HC)14. These contradictory findings could be due to methodological aspects related to differences in the processes studied , paradigms or stimuli used, but also use of small samples, Region of Interest (ROI) analysis or uncorrected statistics. Conversely, heterogeneous findings could be indicative of “real” heterogeneity among BPD patient populations.According to the Diagnostic and Statistical Manual of Mental Disorders 5th edition (DSM 5), Borderline personality disorder (BPD) is characterized by a pervasive pattern of instability referred to interpersonal relationship, self-image and affects together with marked impulsivity and emotional dysregulation15. An earlier meta-analysis on 19 studies16 focused solely on the contrast between negative and neutral stimuli and found higher convergence of differences in BPD as compared to HC in the left amygdalae and in the posterior cingulate, along with a blunted response of the bilateral dorsolateral prefrontal cortex. Here, we implemented a broader approach to BPD as we hypothesized that despite heterogeneity, a consistent pattern of dysfunction in BPD would nonetheless emerge, with more specific patterns observed in homogeneous subgroups. In particular, network analysis highlighted17 difficulties in emotional regulation as central features of BPD, while several reports underscored overlapping brain networks related to negative emotion processes and working memory18 or even a casual effect of difficulties in emotional regulation on other cognitive functions such mentalizing19 and working memory20. Thus, we expected that associated neurobiological dysfunctions would also impact other mental functions and hence emerge consistently across studies, despite the use of different tasks or evaluation domains.Hence, we conducted a systematic review and activation likelihood estimation (ALE) meta-analysis on the fMRI studies conducted in BPD patients compared to healthy controls. The meta-analytical technique considers nuclei of activation reported in single experiments as spatial probability distributions centered at the coordinate itself. These distributions are then used for the generation of a brain map representing the likelihood of activation of each candidate location21 (see The study was pre-registered on PROSPERO repository CRD42019121856) and is reported following the PRISMA guidelines9121856 aFrom each report, we extracted the following information: (1) participant mean age and gender ; (2) comorbidities; (3) concurrent treatments; (5) type of task  and stimuli ; (6) coordinates for direct comparison of brain activation between BPD and HC; (7) where available, coordinates for the activations within each single group (BPD and HC). Data were extracted independently by two researchers .22, (mNOS) adapted to fMRI data23 (See The Risk of Bias (RoB) of included studies were evaluated with a modified version of the Newcastle-Ottawa scale (NOS)24. We used the correction proposed by Turkeltaub and colleagues15, as implemented in GingerALE to control for multiple comparisons on the same dataset. Sample size for each experiment were used to calculate the Full‐Width Half‐Maximum (FWHM) of the Gaussian function used to blur the foci. Coordinates in the MNI 152 standard space were converted into the Talairach space using the GingerALE foci converter tool.Stereotactic coordinates for the ALE meta-analysis were extracted from the studies. The ALE algorithm was used as implemented in GingerALE 3.02 softwareTwo approaches can be employed in an ALE meta-analysis of two groups. The first (“convergence of activation differences”) uses coordinates from the contrast ‘patients vs. controls’ . The second (“differences in convergence”) pools the activation reported within each group separately, and subsequently computes a contrast between the resultant ALE-maps. We performed both for each of the two contrasts of interest  in each experiment.p < 0.001 cluster forming threshold, p < 0.05 cluster corrected FWE and N = 2000 permutations.We used convergence of activation differences as the primary analysis because it used data from all included studies. Statistical significance was assessed and corrected for multiple comparisons using a cluster-based method implemented in GingerALE: To check the robustness of the findings we performed a pooled analysis combining coordinates across BPD > HC and HC > BPD. This analysis reflects a more adequate summary of group differences because at the single study level, differences in analysis approaches and control conditions may have influenced the direction of reported group differences. To explore heterogeneity, we also conducted post-hoc sensitivity analyses by assembling more homogeneous subgroups of similar studies, based on extracted study characteristics such as type of task , type of stimuli  or type of domain evaluated .25 to evaluate the potential publication bias, and a “leave-one-out” analysis to assess the impact of single studies on the results.As additional robustness checks for analyses that produced significant findings, we conducted a “Fail-Safe N” analysis adapted for ALE meta-analysisp value < 0.05, N = 10.000 permutations and a cluster threshold of 100 mm3.For the secondary analysis (differences in convergence), we computed separate meta-analyses for activations of controls and BPD and then contrasted them in a differences of convergence analysis. For the single group meta-analysis, we used the same parameters as in the primary analysis. To compute the differences of convergence, we used an uncorrected Potential differences between the two methods could be attributed to the different number of included studies rather than to genuine discrepancies between methods. We tested for this hypothesis in sensitivity analyses, in which the primary and secondary methods were limited to the experiments reporting coordinates for the same type of contrasts for both single- and between groups results. Fig. .27.Across all analyses, we set the minimum number of studies to 17. According to previous simulations, meta-analyses with less than 17 studies are likely to have insufficient power to detect smaller effects, increasing the risk that results are driven by single/few experimentsn = 9); (2) comparison restricted to functional connectivity analysis (n = 5); (3) non-significant results for the comparison (n = 4); (4) ROI only reported (n = 21); (5) re-analyses of previous, already included, studies or paper reported no new results (n = 1). A total of 52 articles , 371 of which were excluded based on the abstract, leaving 92 reports for full-text inspection. From these, 40 reports were further excluded due to (1) lack of direct univariate comparison between BPD and HC (es Table  were incThe 54 experiments included 2084 subjects (1104 BPD and 1100 HC). All studies performed whole-brain analyses. For the primary analysis we used 52 experiments: 24 reported both contrasts HC > BPD and BPD > HC, seven the HC > BPD contrast only, whereas 21 the BPD > HC contrast only. Twenty-three studies also reported single group analyses Fig. : 21 for According to our mNOS scale scores no study presents a high risk of bias, eleven have low risk while the great majority (41) have an intermediate risk  were considered. For the HC > BPD meta-analysis we included 31 experiments and obtained a minimum cluster size of 624 (MCS) mm³, while for the BPD > HC meta-analysis we included 45 experiments and obtained an MCS of 752 mm³. Across both ALE meta-analyses, we did not find any cluster of significant convergence.3), we found significant clusters in the two amygdalae along with the anterior cingulate (ACC)/middle frontal gyrus (MFG)  for the contrast BPD > HC , showed that the clusters in the right amygdala remained significant 29 over 34 times, the cluster in the left amygdala, 27 times and the ACC/MFG cluster 24 times. Notably, excluding one particular studyrs Table .Table 3LTwenty-three experiments reported coordinates for single group analyses (23 for HC and 22 for BPD) . ResultsThe number of eligible studies for all sensitivity analyses less than 17. Nevertheless, we computed exploratory sensitivity analyses (considering only emotion processing and respectively only active task experiments), which showed no significant results see  Supplem.41, impulsivity42, attention and working memory43), as well as study populations . Importantly, though impulsivity and emotional dysregulation are cardinal symptoms of BPD, they were assessed in few studies (four and six respectively). However, methodological differences can represent other important sources of heterogeneity. A variety of behavioral tasks were used, and it is likely that some were more reliable and robust in measuring the target processes than others, as demonstrated in a large-scale analysis of self-regulation measures44. Moreover, analytic pipelines including pre-processing, choices in data analysis like the type of multiple comparison corrections employed most likely diverged between studies, with direct consequences on the threshold for identifying statistical findings48. The relationship between analytic choices and reporting statistically significant findings is particularly relevant for ALE meta-analyses, which exclusively rely on these results and cannot consider non-significant one49.We report, to our knowledge, the first meta-analysis comprising all fMRI neuroimaging studies for borderline personality disorder. The main findings underscore the substantial heterogeneity of this literature, in reference to the processes studied  reflecting the large number of studies published in the last 4 years. Other sources of discrepancy include different inclusion criteria, meta-analytic approaches  and contrasts selected in the analysis.Our findings partially confirm those of an earlier meta-analysis by Schulze and colleagues25, for findings to be robust to publication bias, at least twice the number of included studies should be necessary to make them non-significant. In the present report, additional evidence for publication bias comes from the high number of studies that could not be included due to not reporting significant findings or conducting only ROI analyses. Thus, the hypothetical null studies suggested by the Fail-Safe N might not even have to be searched in the file-drawer, but already published. Solutions for clarifying similar issues in the neuroimaging literature include access to unthresholded maps, for example by posting them in public repositories, so as to allow studies with null or negative findings to be included in neuroimaging meta-analyses. Likewise, pre-registration of ROI analyses49, by guaranteeing these analyses were not contingent to non-significant whole-brain results, could support their inclusion in meta-analyses. The results of the leave-one-out analysis for the emotional processing subgroup similarly point to limited robustness. The three significant clusters were maintained in only 13 iterations, but with the exclusion of a single study37, no results remain significant. This suggests that findings are heavily influenced by a few individual studies.Despite the considerably larger pool of studies, the robustness of the emotional processing findings is limited. Fail-Safe N analysis showed that as few as three additional null studies would render the clusters in the ACC and in the left amygdala no longer significant. Conversely, for the right amygdala, 33 studies with null findings would render the result not statistically significant. The Fail-Safe N is a proxy for potential publication bias . According to the adaptation of this method for ALE meta-analysisFinally, we failed to find convergence for the HC > BPD. While this contrast was reported in only 24 experiments, it is interesting to note that only four studies specifically discussed the dysfunction of emotional processing. As dysfunctions in emotional modulation are believed to be caused by a lack of activation in the prefrontal cortex in BPD, we speculate that a higher number of studies on emotional regulation would result in significant HC > BPD convergence.18, thus is not surprising how control subjects showed a higher convergence in this region as compared to BPD. This might be due to a general lack of IFG activity in BPD or to a higher heterogeneity of brain activations in this group.The contrast HC > BPD resulted in a significant cluster of difference of convergence in the right insula/IFG complex. The activation of the inferior frontal gyrus has been often described for tasks related to emotional regulation and modulation26. The former aims at identifying common difference between groups across the experiments, whereas the latter focuses to the convergence within the groups across the experiments and then to their possible differences. We confirm that these two approaches produce divergent findings, as previously shown70. Comparison of the two methods restricted to experiments reporting the data required to run both resulted in a significant cluster in the anterior cingulate for convergence of difference, and another in the left insula for difference of convergence, for the HC > BPD contrast, suggesting that divergences between the two methods are not imputable to different numbers of studies.According to Muller and colleagues the difference between the convergence of difference and difference in convergence approaches are mirrored by the research question one wants to answer70. In our work, differences might also be related to the relatively poor robustness of the primary analysis findings. For example, Hazellet and colleagues8, a study that if excluded resulted into two of the three significant clusters becoming non-significant, did not provide single group results, so could not be used for the difference in convergence analysis.The discrepancy between the two approaches is probably rooted into the ALE meta-analysis method of taking only coordinates of significant clusters. We proposed that the convergence of differences method is more feasible for contrasting groups, because it generally relies on a larger number of studies and is usually more concordant with findings from individual studies71: the depressed groups of each study might be too neurobiologically different among each other thus the convergence of difference did not produce consistent results. Specifically, while we found no significant clusters with the difference of convergence method, we did find clusters of convergence for both HC and BPD with the convergence of difference method. This suggests that despite task-related heterogeneity, a common pattern of activation within BPD patients can be identified. We suggest that this result is a hint to the fact that the heterogeneity of BPD population is similar to the one of healthy controls although a more formalized and quantitative approach to heterogeneity should be applied  to confirm this speculation.However, the two methods could also be viewed as complementary. In a recent ALE meta-analysis on depression the heterogeneity of the depressed patients was proposed as a possible reason for the lack of significant convergence24. Moreover, coordinate based meta-analyses rely exclusively on coordinates for contrasts that reached statistical significance, as defined in each individual study. Consequently, these methodologies involve a significant loss information as compared to aggregation of fMRI unthresholded maps73 and cannot take into account publication bias25. Unfortunately, open sharing of unthresholded maps remains limited74. Another limitation is related to the considerable number of studies that were excluded due to reporting only on ROI analyses (N = 23). Though the ROI approach is widely employed in neuroimaging studies, particularly in small sample studies, it is associated with increased risk of false positives or overestimation of effects49. Moreover, its use is discouraged in coordinate based meta-analysis26. Unfortunately, requests for whole-brain results are often not honored54.By design, ALE meta-analysis can only quantify convergence probabilities and not magnitude of activations. The method provides a probability of convergence, that is, concordance of statistically significant foci across experiments in terms of probability distributions centered at the each set of focus coordinates71. Though this meta-analysis included a large number of neuroimaging experiments (99), overall analyses across cognitive and emotional processing experiments, as well as subgroup analyses, revealed no consistent pattern of convergence. As possible causes for the lack of significant convergence of activation, the authors list differences among individual studies  and meta-analytic approaches . Though we did report a pattern of amygdala and ACC dysfunction in studies of emotion processing, it was not robust to possible publication bias and single study effects. Amygdala dysfunction might represent a promising pathophysiological mechanism, as a well target for novel therapeutic strategies65, though, conversely, it might also indicate a transdiagnostic marker of difficulties in emotional processing and regulation. Further studies may disentangle this aspect particularly evaluating emotional regulation difficulties both in BPD and across psychopathological domains with a single paradigm. Network analyses have shown how difficulties in emotional regulation are the most relevant core feature of BPD17, applying the same approach to neuroimaging data across psychiatric disorders may help to clarify the specificity of the present findings. Perhaps more importantly than conducting new studies, public sharing of unthresholded maps75 from already conducted ones would represent a great advance in the understanding of the neurobiology of BPD.A meta-analysis of neuroimaging studies identified a consistent, albeit not robust, pattern of activation for emotional processing. No other common pattern of convergence of activation emerged, probably owing to the high between-study heterogeneity, ranging from tasks, populations and analytic approaches. Our findings mirror those of another ALE meta-analysis on unipolar depressionSupplementary material


However, the results were not consistent. We therefore conducted a meta‐analysis to explore the underlying relationships between BDNF rs6265 G > A polymorphism and PTSD risk.Previous studies have shown that Five online databases were searched, and all related studies were reviewed up to July 1, 2020. Odds ratios (ORs) and corresponding 95% confidence intervals (CIs) were calculated to examine the statistical power of each genetic model. In addition, heterogeneity, sensitivity accumulative analysis, and publication bias were examined to check the statistical power.Overall, 16 publications involving 5,369 subjects were included in this systematic review and 11 case‐control studies were analyses in meta‐analysis. The pooled results indicated an increasing risk of A allele mutations with PTSD risk. Moreover, the sequential subgroup analysis also demonstrated some similar situations in Asian populations and other groups.BDNF rs6265 G > A polymorphism might be involved in PTSD susceptibility.Current meta‐analysis suggests that the  A meta‐analysis focused on the association between BDNF rs6265 G > A polymorphism and PTSD risk was conducted. 16 publications involving 5,369 subjects were included in this system review and 11 case‐controls were analyses in meta‐analysis. Current meta‐analysis suggests indicated that the BDNF rs6265 G > A polymorphism might be involved in PTSD susceptibility. In the In recent decades, there has been increasing evidence that PTSD is caused by interactions between various neural and traumatic factors  locus in the 5′ promoter region of the BDNF gene, which is located on the short arm of chromosome 11p13. This gene variant involves a nucleotide substitution from guanine to adenine at position 196 in the BDNF coding region, resulting in a nonsynonymous amino acid alternation from valine  to methionine (Met) in codon 66 of the BDNF prodomain  statement , and two Chinese databases (CNKI and Wanfang) were used to identify studies on the association between BDNF rs6265 G > A polymorphism and PTSD susceptibility from database inception to July 1, 2020. The bibliographies of all included studies were reviewed to identify additional relevant studies. The strategy was listed :#1 Brain Derived Neurotrophic Factor#2 BDNF#3 rs6265#4 #1 OR #2 OR #3#5 polymorphism#6 variant#7 mutation#8 #5 OR #6 OR #7#9 Post‐Traumatic Stress Disorder#10 PTSD#11 #9 OR #10#12 #4 AND #8 AND #112.2The following criteria were used to identify relevant studies: (a) only case‐control and cohort studies were selected; (b) studies on the association between BDNF rs6265 G > A polymorphism and PTSD susceptibility were selected, and subsequent meta‐analyses were conducted with studies in which the P value of the Hardy–Weinberg equilibrium (HWE) test in genotype distributions in the control group was greater than 0.05; (c) studies with sufficient genotype data for both case and control groups included; (d) studies published in English or Chinese; (e) subgroup analyses were conducted with at least two groups; and (f) studies with the latest or largest sample size were retained if multiple publications or overlapping data were found. The exclusion criteria included the following: (a) case report (case series), review articles; (b) biological fundamental and animal experiment studies; and (c) studies without sufficient genotype information.2.3−]), genotyping method, sample sizes of cases and controls, frequency information for the genotype distribution of the case and control groups, traumatic factors, assessment of the HWE in the control group, age distribution, and diagnostic criteria.Two authors (Hu and Wu) independently reviewed all included studies and extracted the following information: the name of the first author, publication year, country and subject ethnicity, control design  was considered statistically significant.Crude odds ratios (ORs) and 95% confidence intervals (CIs) were calculated to examine the association between the BDNF rs6265 G > A polymorphism and PTSD susceptibility. Five genetic models of the rs6265 G > A polymorphism were examined: allele contrast (A versus. G), co‐dominant (GA versus. GG and AA versus. GG), dominant (GA + AA versus. GG), and recessive (AA versus. GG + GA). Heterogeneity among the included studies was examined using Cochran's Q and I33.1The selection process is illustrated in Figure 3.2p =.02, I2 = 18.5%; AA versus. GG: OR = 1.46, 95% CI = 1.11–1.92, p =.01, I2 = 19.8%,    . A similar increased risk of rs6265 G > A mutation in patients with PTSD risk was also observed following natural disasters and diseases exposure  , Li Guo et al. , and Guo et al.  .Publication bias was evaluated, and funnel plots did not demonstrate any significant asymmetry  and PTSD . In their meta‐analysis, nine studies were included, but the results obtained were based on different genetic models. Moreover, only one genetic model was examined to determine the association between BDNF rs6265 G > A polymorphism and PTSD susceptibility, without any subgroup analysis or quantitative assessment  More studies were included than in previous analyses; (b) all five genetic models were examined in general and subgroups to explore the potential relationship between rs6265 G > A polymorphism and PTSD; (c) the statistics examined were taken from studies that satisfied the HWE for the genotype distribution in controls; (d) no significant heterogeneity was found, indicating a fair consistency among all included studies; and (e) no significant publication bias was found using Egger's test and Begg's funnel plots.5Our results from this meta‐analysis suggest that the BDNF rs6265 G > A polymorphism is associated with PTSD susceptibility in Asian people. Further studies are needed to assess the association between this SNP and PTSD.The authors declare no competing financial interests.HXY, WYL, LXX, and ZL conceived the study. HXY and WYL searched the databases and extracted the data. HXY and CCH analyzed the data. HXY, LXX, and ZL wrote the draft of the paper. All the authors approved the final manuscript.All collected information was obtained from all published articles, and no ethical approval was necessary.https://publons.com/publon/10.1002/brb3.2118.The peer review history for this article is available at


Background: Insulin-like growth factor-1 (IGF-1) has been demonstrated to increase fatty acid β oxidation during fasting, and play an important role in regulating lipid metabolism and type 2 diabetes mellitus (T2DM). The rs35767 (T > C) polymorphism, a functional SNP was found in IGF-1 promoter, which may directly affect IGF-1 expression. However, the inconsistent findings showed on the IGF-1 rs35767 polymorphism and T2DM risk.Methods: We performed a comprehensive meta-analysis to estimate the association between the IGF-1 rs35767 and T2DM risk among four genetic models .Results: A total 49,587 T2DM cases and 97,906 NDM controls were included in the allele model, a total 2256 T2DM cases and 2228 NDM controls were included in the other three genetic models . In overall analysis, the IGF-1 rs35767 was shown to be significantly associated with increased T2DM risk for the allele model , additive model  and dominant model  with random effects model. After omitting Gouda’s study could reduce the heterogeneity, especially in the recessive model , the fixed effects model for recessive effect of the T allele (TT vs. CC + CT) produce results that were of borderline statistical significance . And increasing the risk of T2DM in Uyghur population of subgroup for the allele model.Conclusion: The initial analyses that included all studies showed statistically significant associations between the rs35767 SNP and type 2 diabetes, but after removing the Gouda et al. study produced results that were mostly not statistically significant. Therefore, there is not enough evidence from the results of the meta-analysis to indicate that the rs35767 SNP has a statistically significant association with type 2 diabetes. Substanβ) cells . Insulinβ) cells . Studiesβ) cells .Insulin-like growth factor-1 (IGF-1) is a circulating growth factor which structure is highly homologous with pro-insulin. IGF-1 is expresses in insulin-resistant tissue, it downregulates free fatty acid and increases fatty acid β oxidation during fasting (IGF-1 level are prone to have diabetes mellitus (IGF-1 gene can directly affect IGF-1 expression. The rs35767 (T > C) polymorphism, a functional SNP was found in IGF-1 promoter, in which the promoter with C allele showed a higher transcriptional activity than promoter with T allele  or rs35767 (A > G),” “polymorphism or SNP or mutation or variant” and “diabetes or type 2 diabetes or T2DM.” All searches had no language limitations. Eligible studies were estimated by reading full texts, and excluded substandard studies.The Google Scholar, PubMed and Chinese National Knowledge Infrastructure were comprehensively searched for related studies published before July 31, 2021, using the key terms: “insulin-like growth factor 1 or IGF-1 rs35767 and T2DM risk; 2) sufficient raw data or adequate data for assessing odds ratios (ORs) with corresponding 95% confidence intervals (CIs); and 3) The diagnostic standard of T2DM conformed to the World Health Organization.The following inclusion criteria: 1) case-control or cohort studies that relate to the IGF-1 rs35767 and T2DM risk; 3) lacking detailed data; and 4) control subjects is not in Hardy-Weinberg equilibrium (HWE).The following exclusion criteria: 1) not a case-control study; 2) irrelevant to 2), the distributions number of genotype and alleles, ORs with 95% CI, or ability to calculate the OR and 95% CI. p-value for the HWE of NDM controls.Data were independently extracted by two authors from the eligible studies and collected the following data: first author, year of publication, origin, the numbers of T2DM cases and NDM controls, gender and age, BMI . Four genetic models were evaluated in this meta-analysis: the allele model (T vs. C); the additive model ; the recessive model (TT vs. CC + CT) and the dominant model (TT + CT vs. CC). Using Q-test and II2 > 50% . OtherwiI2 > 50% .A total of 182 potential articles obtained through initial search. 51 duplicates were excluded. Then 131 studies were screened on title and abstract, 84 of them were excluded. The left 47 articles were evaluated by full-text reading, 35 of them were excluded cause that 22 were not case-control researchs, 10 were not related to rs35767 or T2DM, three did not provided sufficient data. 12 articles were included that there are six articles including five in English and one in Chinese just of the allele model data, and other six articles including four in English and two in Chinese of four genetic models data . Flow chart of researches selection in the meta-analysis was shown in IGF-1 rs35767 polymorphism and T2DM were evaluated using ORs and 95% CI in the allele model  and the additive; recessive and dominant models .The association between the IGF-1 rs35767 was shown to be significantly associated with increased T2DM risk for the allele model , additive model  and dominant model . The results showed no significant difference for the recessive model   .2 = 38.7%, p = 0.163), the fixed effects model for recessive effect of the T allele (TT vs. CC + CT) produce results that were of borderline statistical significance ; In addtion, other three modal show moderate degree of heterogeneity , the result showed no significant association between IGF-1 rs35767 and T2DM risk with random effects model   . Since tXinJiang, China subgroup  with fixed effects model; a random effects model were used to analyze the other provinces, China, rs35767 was shown no significant association with T2DM risk ; and not a significantly associatied in the other countries subgroup   with fixed effects model  . Then omIGF-1 rs35767 and T2DM risk, only six articles had four gene models data, and the other six articles had only one allele model data. Secondly, the results did not adjustment the potential risk factors, including gender, body mass index, age, drinking and smoking status, and environmental factors. Thirdly, some results showed relatively obvious heterogeneity, but research the source of heterogeneity needs to more larger sample. Finally, some groups results existed potential publication bias in Egger’s test. Therefore, the results of the article should be interpreted carefully.There still have several limitations in our meta-analysis. Firstly, there were limited studies which estimated IGF-1 rs35767 and T2DM susceptibility. Overall, there is not enough evidence from the results of the meta-analysis to indicate that the rs35767 SNP has a statistically significant association with T2DM. Further more studies are necessary to verify the results.This is the first time to perform a meta-analysis to systematically summarize the association between the


Begg’s funnel plot and Egger’s linear regression test were used to evaluating possible publication bias among the included studies, and sensitivity analysis was used to test the stability of the overall results. A total of 45 studies met our selection criteria and eight related genetic association studies were retrieved, including 320 single-nucleotide polymorphisms from 20 candidate pathways, ranging from 2000 to 2021. A total of 28,994 healthy people versus 20,600 IgA nephropathy patients were enrolled. Upon meta-analyzed results that TGFB1 , IL-1B (rs1143627), IL-18 (rs1946518), and TLR1 (rs5743557) showed effect with or without ethnicity difference. And 10 variants presented stable and robust related to IgA nephropathy. This research showed that genetic variants are related to the immunologic and inflammatory effects of IgA nephropathy pathogenesis. The meta-analysis results supported the previous researches, and may help deepen the understanding of pathogenesis and explore new targets for IgA nephropathy-specific immunotherapy.IgA nephropathy is the most prevalent primary glomerulonephritis worldwide, with identical immunopathological characteristics caused by multiple etiologies as well as influenced by geographical and ethnical factors. To elucidate the role of immunologic and inflammatory mechanisms in the susceptibility to IgA nephropathy, we explored single nucleotide polymorphisms of related molecules in the immune pathways. We searched the PubMed database for studies that involved all gene variants of molecules in the 20 immunologic and inflammatory pathways selected from the Kyoto Encyclopedia of Genes and Genomes database. The odds ratios with their corresponding 95% confidence intervals in six genetic models  were summarized using fixed or random effect models. Subgroup analysis was conducted based on different ethnicities with generalized odds ratios. Heterogeneity was evaluated using the Q and I G) analyzes the account of cases/healthy control pairs in the study to indicate the mutational load of disease susceptibility (IgA nephropathy (IgAN) is the most prevalent primary glomerulonephritis in the world, and 30–40% of patients progress to the end-stage renal disease within 20–30 years after diagnosis, requiring kidney transplantation or renal replacement therapy . Howevertibility .G. However, the large disparities in prevalence based on ethnicity that most studies have reported in cohorts of Asian populations has limited the overall review and meta-analysis of IgAN.In past studies, the chosen of genetic models might not have been detailed enough to support results. In order to provide a strong support to existing evidence, we performed cluster analysis on six genetic models using SNPs in studies and calculated ORTo elucidate the role of the immunologic and inflammatory mechanisms in the susceptibility to IgA nephropathy, we selected 20 candidate pathways classified by the Kyoto Encyclopedia of Genes and Genomes (KEGG) database  the IgA nephropathy diagnosis was confirmed by renal biopsy; (2) healthy individuals included in the study as a control group; (3) sufficient genotypic data, such as genotype count data or allele frequencies, were provided; and (4) disease susceptibility was investigated.in vitro experiments; and (6) data unextractable studies.Studies meeting any one of the following criteria were excluded: (1) the research subjects were patients after kidney transplantation or recurrent patients; (2) no healthy subjects as a control group; (3) studies investigating progression, prognosis, severity, response to treatment; (4) comments, case reports, editorials, letters, reviews, and non-English articles; (5) animal studies and Two reviewers independently screened the abstracts of the retrieved articles, and if necessary, the full text was also reviewed. Discrepancies regarding potentially eligible studies were finalized through discussion to reach a consensus.We used electronic data to extract the following information from each eligible study: first author, publication year, study country, PMID, ethnicity, disease type, sample size, age, and sex. For genotypic data, we extracted genotype count data in cases and controls, and, if available, the odds ratios (ORs) with their corresponding 95% confidence intervals (CIs) of five genetic models and evidence of Hardy-Weinberg equilibrium (HWE) in controls were also recorded.We appraised the quality and risk of bias of the included studies according to the Newcastle–Ottawa Quality Assessment Scale (NOS). The total score of NOS is 9 points, including three aspects: selection, comparability, and exposure. According to the final score, the studies could be categorized into high quality (score >6), medium quality (score between 4 and 6), and low quality (score less than 4).2≥50% or p < 0.05 of the Q test, indicating that the homogeneity was insignificant, then the random-effects model (DerSimonian and Laird method) was considered. Since genetic models are not independent and there is no a priori biological justification for the choice, we performed genetic association analyses using six genetic models  for each molecule  software that was programmed by statist Elias Zintzaras (STATA SE software (version 15.0) was used to conduct all the data processing. HWE was examined using Fisher’s exact test and whether controls were confronted with HWE according to P-value results, with 0.05 as the boundary. The association between the mutational load of a variant and susceptibility to IgAN was presented by ORs with 95% CIs. For each variant, the OR estimates were calculated using the fixed-effects model (Mantel-Haenszel method). Statistical significance was set at P < 0.05. If the heterogeneity test Imolecule . Multiplmolecule . Hence, intzaras .Additionally, when the heterogeneity between studies was statistically significant, we used meta-regression analysis to identify potential sources of such heterogeneity. Begg’s funnel plot and Egger’s linear regression test were used to evaluating possible publication bias among the included studies. The stability of the combined results was examined by performing a sensitivity analysis.The workflow of the literature selected for meta-analysis is shown in core 1 synthase, glycoprotein-n-acetylgalactosamine 3-beta-galactosyltransferase 1 (C1GALT1), complement factor B (CFB), Cytotoxic T-Lymphocyte Associated Protein 4 (CTLA4), C-X-C Motif Chemokine Ligand 8 (CXCL8), endothelial nitric oxide synthase (Enos), Fc Receptor Like A (FCRLA), FCRLB, G Protein Subunit Gamma 2 (GNG2), Major Histocompatibility Complex, Class II, DP (HLA-DP), Inducible T Cell Costimulator (ICOS), Interferon-γ (IFN-γ), Interleukin 10 (IL-10), IL-1B, Interleukin 1 Receptor Antagonist (IL1RN), IL-6, Integrin Subunit Alpha X (ITGAX), LEM Domain Nuclear Envelope Protein 2 (LEMD2), Netrin 4 (NTN4), Pleckstrin Homology Like Domain Family B Member 1 (PHLDB1), Signal Transducer And Activator Of Transcription 4 (STAT4), Transforming Growth Factor-β (TGF‐β1), Toll Like Receptor 1 (TLR1), TLR10, Tumor Necrosis Factor Superfamily Member 13 (TNFSF13), Tensin 3 (TNS3), and Vascular Endothelial Growth Factor (VEGF); the details are shown in We conducted a meta-analysis of all molecules with gene polymorphisms that were confronted with HWE. The results show significant association with susceptivity to IgAN under the allele contrast model, as follows: FCRLB  have all been shown to have a statistical impact on IgAN susceptibility in the allele model CTLA4 has three SNPs , all of which were significant. Both rs9277535 and rs3077 in HLA-DP showed an increased risk of developing IgAN. In addition, ICOS has eight gene variants, five of which have protective effects under the allele model  were associated with susceptibility to IgA in It is worth mentioning that the five gene variants of C1GALT1, CFB, CTLA4, CXCL8, FCRLA, FCRLB, HLA-DP, ICOS, IFN-γ, IL-10, IL-1β, IL1RN, IL4R, IL5RA, IL-6, ITGAM, ITGAX, Netrin 4 (NTN4), ST6GAL1, ST6GALNAC2, TGF‐β1, TLR1, TLR10, TNFRSF6B, TNFSF13, and TNS3. The general characteristics and details of each research on significant SNPs are described in ICOS has five SNPs , all of which showed a robust significant risk of developing IgAN under dominant model which corresponded with the allele model results. Besides that, TNFRSF6B with the mutations of rs1291205 [OR = 0.45 (95% CI 0.31–0.54)], rs1291206 [OR = 0.46 (95% CI 0.31–0.67)], rs3208008 [OR = 0.47 (95% CI 0.32–0.69)] showed significant protective effect. And we obtained interesting results that did not show significance in allele model. The SNPs rs12054151, rs2239611, rs1990677, rs4686838, rs2284750, rs6784233, and rs7634389 in ST6GAL1 showed robust association with the occurrence of IgAN; rs3840858 in ST6GALNAC2 [OR = 3.68 95% CI (1.75–7.73)] showed a great correlation with IgAN risk in Chinese Uyghur population.To further validate the association of these significant SNPs with the risk of IgAN under allele model, we analyzed all gene polymorphisms in dominant contrast model. The genes significantly associated with IgAN occurrence and compounded with HWE were as follows: ST6GALNAC2 showed protective effect respectively with [OR = 0.27 (95% CI 0.13–0.57)] and [OR = 0.64 (95% CI 0.41–0.99)] under the overdominance model which contradicted with dominance model result, but still hinted at IgAN occurrence in Uyghur Chinese population. The five SNPs of ICOS mentioned above were stably and significantly associated with IgAN but were associated with protective effects under overdominance model. Our meta-analysis results showed that rs1203350 and rs1203344 in IGAN1 decrease the susceptibility of IgAN in Helsinki and Toronto population, but not in St. Etienne patients. Besides that, according to the results of overdominance model, rs6569686 decreases the susceptibility of IgAN in Helsinki people ], dominant model [OR = 0.68 95% CI (0.51–0.90)], and overdominant model [OR = 1.39 95% CI (1.04–1.85)]. Several SNPs showed significance only in homozygote model compared to the other genetic models mentioned above including Activating Transcription Factor 6 (ATF6), Bone Morphogenetic Proteins 2 (BMP2), C Motif Chemokine Receptor 6 (CCR6), F-Box And Leucine Rich Repeat Protein 21 (FBXL21), LEM Domain Nuclear Envelope Protein 2 (LEMD2), Pecanex 3 (PCNXL3), RAS Guanyl Releasing Protein 1(RASGRP1),and Regulator Of G Protein Signaling 1 (RGS1). The details of the homozygote model with p<0.05 and HWE≥ 0.05 are shown in IL-1B under homozygote model are shown in It is generally recognized that homozygote of the mutation increases the risk for disease than the heterozygote, whereas homozygotes of the wild allele are not risky . C1GALT1HLA-DP rs3077 [OR = 0.31 (95% CI 0.18–0.52)], STAT4 rs7574865 [OR = 0.56 (95% CI 0.35–0.89)], TGF‐β1 rs6957 [OR = 0.16 (95% CI 0.04–0.64)] and TGF‐β1 rs180047 [OR = 0.04 (95% CI 0–0.87)], TLR1 rs4833095 [OR = 2.13 (95% CI 1.22–3.71)], TNS3 rs3750163 [OR = 17.99 (95% CI 4.27–75.71)]; all had a significant effect which was identified as having an OR greater than 1.5 or less than 0.5, the results shown in CFB, C1GALT1, and CCR6, CD29, CTLA4, CXCL8, F-Box and Leucine Rich Repeat Protein 21 (FBXL21), Fc Fragment Of IgG Receptor IIb (FCGR2B), FCRLA, FCRLB, HLA-DP, ICOS, IFN-γ, IL-10, IL-1β, ITGAM, ITGAX, RGS1, STAT4, TGF‐β1, TLR1, TLR10, TNFSF13, and TNS3. General characteristics of each studies about molecules are described in Under heterozygote model, compared to healthy controls, ICOS showed significant effect again with the mutations of rs4404254, rs10183087, rs11571314, rs1559931, rs4270326, and IL-10 rs1800872, CXCL8 rs2227543, CXCL8 rs2227306, ITGAM rs4597342, and TGF‐β1 rs2241715) have a significant effect in dominant model, overdominant model, and heterozygote model. Among alleles, FCRLA rs1954174, IL1RN rs928940, IL1RN rs315951, LEMD2 rs751728, and VEGF 405C-G showed significance in homozygote and recessive models. Furthermore, HLA-DP rs3077 and STAT4 rs7574865 showed statistical significance in allele model, overdominant model, homozygote model, heterozygote model, and recessive model. In addition, TNFSF13 rs3803800 showed the most stable effect under five genetic models excepted the overdominant model; CFB rs549182 has a robust impact on susceptibility in all models except the homozygote one; there were two gene variants of CFB (rs549182 and rs4151657) that were significant in different patient groups with IgA nephropathy and we summarized it with ORG in All results of recessive model were described in C1GALT1, HLA-DP, ITGAX, ITGAM, TNFSF13, and ST6GAL1 showed significant associations according to our meta-analysis results, further confirmed by GWAS studies. Recently, a large-scale genome-wide meta-analysis included four GWAS studies on people of Chinese and European descent to investigate genetic variants of IgAN susceptibility , interferon regulatory factor 4 (IRF4) rs6942325, and Peptidyl Arginine Deiminase 4 (PADI4) rs2240335.To date, eight GWASs have been performed that provide valuable information on the susceptibility of IgAN. tibility . This anG and 95% CI. We summarized published studies related to rs1800469, rs1982073, and rs1800471 in TGFB1. Excluding those that did not meet our selection criteria, we finally enrolled five studies involving Korean, Japanese, Italian, and Swedish populations  (G = 0.68 95% CI (0.63–0.74)], rs9275424 [ORG = 1.26 95% CI (1.09–1.42)]; rs2856717 [ORG = 0.76 95% CI (0.69–0.83)], and rs9275596 [ORG = 1.12 95% CI (1.04–1.19)] had consistent significant results in European and Asian cohorts  mutation; it showed a significant protective effect with overall OR46518 mut39–0.73)  (Supple39–0.73) , 41 (Su39–0.73) , 43. BesIn the sensitivity analysis, recalculated ORs after removing any single eligible study showed that no material alterations were detected implying that our results are reliable and robust . These studies were consistent with the global prevalence of IgAN. Although the prevalence in some Western countries has progressively increased, they still have a lower prevalence of IgAN, and Asian populations such as the Chinese, Japanese, and Korean are more vulnerable to IgAN .S) are formed, (4) of which these ICS accumulated in the glomerular mesangium, induce cell proliferation, and secrete immune molecular to aggravate renal injuries  aberrant glycosylation of IgA1. (2) anti-glycan/glycopeptides antibodies against unnormal IgA1 synthesized (3) aberrant glycosylation of IgA1 combined with anti- glycan/glycopeptides antibodies so that immune complexes , αNacetylgalactosaminide α2,6sialyltransferase 2 (encoded by ST6GALNAC2), and core 1 β1,3galactosyltransferase-specific chaperone 1 (encoded by C1GALT1C1)  . Two stuty of it , 25. Acc to IgAN .Upon investigation of 20 immune pathways, we obtained results consistent with previous research that several significant SNPs were involved in the complement pathway especially the alternative pathway. CFB plays a pivotal role in the production of C3; it combines with C3b to form C3 convertase in the complement response . ProductTGF-β1 encoded by TGFB1 is a multifunctional cytokine that contributes to cell proliferation, differentiation, expansion of mesangial matrix, the development of fibrosis, and various immunologic processes  are a family of innate immune receptors expressed in the membrane of leukocytes, renal tubular epithelial cells, and various cells which induce inflammatory cytokine expression through intracellular signaling pathway. The agonized TLRs can directly injure kidney and also overproduce antibodies phocytes . TLR1 iss grades .Interleukin-1 receptors shared the same pathway with TLRs to produce cytokines. IL-1Ra is an anti-inflammatory contributor, while IL-1β as pro-inflammatory, which balances the immune response. In patients with IgAN with a lower IL-1Ra/IL-1βr ratio, more severe pathological changes were observed . IL-1B .ST6GALNAC2 gene and decreased the activity of core 1 β1,3galactosyltransferase 1 encoded by C1GALT1, leading to the accentuation of aberrant glycan-deficient IgA  and the National Key R&D Program of China (Nos. 2016YFC1305500).The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.


Backgound: The high incidence of thiopurine-induced myelosuppression in Asians is known to be attributable to genetic variation in thiopurine metabolism. A quantitative synthesis to summarize the genetic association with thiopurine-induced myelosuppression in Asians was therefore conducted.Methods: A Literature search was performed from January 2016 to May 2021 in the following databases: PubMed, Web of Science, and Embase and addition search included the studies from Zhang et al. Two reviewers independently extracted the following data: the author’s name, year of publication, ethnicity, drugs, diseases, genetic polymorphisms, onset, type of myelosuppression and results of Hardy-Weinberg equilibrium. The Newcastle-Ottawa Scale was used to assess the quality of the studies. The pooled odds ratios (OR) and 95% confidence intervals (CI) were calculated to evaluate the associations of NUDT15 and the risk of thiopurine-induced myelosuppression stratified by onset and type of myelosuppressive. Subgroup analysis by NUDT15 genetic polymorphisms was performed.Results: A total of 30 studies was included in this meta-analysis. The overall OR for the relationship between NUDT15 genetic polymorphisms and thiopurine-induced early onset of leukopenia and neutropenia in Asian populations were 11.43 (95% CI 7.11–18.35) and 16.35 (95% CI 10.20–26.22). Among NUDT15 polymorphisms, NUDT15*3 showed a significantly increased risk of early leukopenia  and early neutropenia . A significantly higher thiopurine-induced early neutropenic risk was also found for NUDT15*2 . Whereas, NUDT15*5 and NUDT15*6 variants showed a lower risk of leukopenia.Conclusion: This study suggests that NUDT15*3 and NUDT15*2 are important genetic markers of thiopurine-induced early onset of myelotoxicity in Asians, therefore, early detection of these variants before initiating thiopurine therapy is necessary. Thiopurines, 6-mercaptopurine (6-MP) and azathioprine (AZA), are purine analogs that are closely related in their structures . 6-MP isTPMT polymorphisms result in TPMT deficiency thereby increasing concentration of 6-thioguanine nucleotide (6-TGN) levels related to myelosuppression  by Thiopurine S-methyltransferase (TPMT) . It is wpression . The fren Asians . Recentln Asians . The disucasians . NUDT15 ucasians  thereforpression .NUDT15*3 variant and thiopurine-induced myelotoxicity is well recognized, there is still controversy about the increased risk of thiopurine hematotoxicity in patients who carry other variants of this gene, particularly NUDT15*2, NUDT15*5, NUDT15*6 which exist in high allele frequencies in Asian populations  the study population was of patients treated with thiopurine drugs: 6-MP, AZA or 6-TG 2) Studied the association between genetic polymorphisms of NUDT15 and thiopurine-induced myelosuppression in an Asian population 3) the outcomes of interest included myelosuppression  4) the study provided sufficient information to calculate the genetic association with thiopurine-induced myelosuppression. Exclusion criteria are 1) studies not relevant to pharmacogenetic of NUDT15 and TPMT and thiopurine-induced toxicity in Asian; 2) not clinical study; 3) review study, systematic review, meta-analysis, letters, editorials, opinion, commentaries, case report; 4) no full text available; 5) not report odds ratio or no sufficient information for calculate odd ratio or data not related to dose reduction. Any disagreements were discussed until consensus between the two reviewers could be reached.Two reviewers (KK and WT) independently assessed abstracts and titles retrieved from the comprehensive searches for study inclusion. Articles from the updated search and from Zhang et alData extraction was performed by two independent reviewers. Any disagreement was discussed and the data checked again to arrive at an agreement. The following data were extracted from each study: the first author’s last name, year of publication, ethnicity, drugs used, disease type, genetic polymorphisms, onset and type of myelosuppression. The Hardy-Weinberg equilibrium (HWE) was tested to check if the included individuals were in equilibrium for the frequencies of genotypes . EquilibNUDT15 variants. Statistical heterogeneity was assessed via the Q statistic and I2 tests  and 95% CI were calculated to determine the association between genetic polymorphisms and the risk of thiopurine-induced myelosuppression. All analyses were performed with the method by DerSimonian and Laird  using a I2 tests . p ≤ 0.1 studies . The Fun studies . All anaA total of 431 studies were identified from an updated literature search while an additional seven studies from a previous systematic meta-analysis by Characteristics of 30 studies are described in 2/d, while the studies conducted in IBD or autoimmune diseases was AZA 0.5–3 mg/kg/d. The distribution of observed alleles and expected alleles of each genetic variation were consistent with HWE, except for three studies  with early onset and 281 cases (40.67%) with late onset carried the NUDT15*3 variant. The higher risk for development of early leukopenia was found to be significantly associated with NUDT15*3 carriers  (2 = 47.4)  (In thirty studies with = 63.6%)  compared = 47.4) . In addi= 0.356) .NUDT15*5 and NUDT15*6 with the risk of myelosuppression. Overall, with those who carried either NUDT15*5 or NUDT15*6, 15.12% (26 of 172) cases had early leukopenia whereas 10.70% (32 of 299) cases had late leukopenia. For neutropenia, 17.39% (4 of 23) cases with early onset and 14.55% (8 of 55) cases with late onset carried NUDT15*6. The OR for early and late leukopenia was about 3.01–4.9 for NUDT15*5 and NUDT15*6 variants   (Six studies were included to determine the association of variants . Only on8–40.89) . There w–708.69)  but not –708.69) .p > 0.05, see Begg’s and Egger’s test were performed to assess for small study effect. The Egger’s test and Begg’s test were not significant for all analysis , NUDT15*3 , NUDT15*5 , and NUDT15*6   and 16.35 (95% CI 10.20–26.22). The higher risk was noted in patients who carried NUDT15*3 more than any other variant with an OR of 15.31 for early leukopenia and 15.85 for early neutropenia. In addition, an almost 38-fold increase of risk of early neutropenia was also found for NUDT15*2 carrier patients.Recently, a number of studies suggested that n Asians . To datereported . The comsGGAGTC)  in whichulations . This cuNUDT15*3 and NUDT15*2 showed a 100% loss of enzyme activity  (. A similar result was reported in that the NUDT15*2 variant showed an approximate 38-fold higher risk of early neutropenia in ALL patients who were treated with 6-MP (activity . The recbolizers . It has 232 SNP) . A similith 6-MP . This suith 6-MP .NUDT15*3 and NUDT15*2, an in vitro study reported that NUDT15*5 and NUDT15*6 showed a loss of enzyme activity of about 50–60%, however, in vivo activities of these enzymes were not quite clear (NUDT15*6 (NUDT15*6 and NUDT15*5 compared to NUDT15*3 (NUDT15*5 or NUDT15*6 with the risk of myelosuppression compared to NUDT15*3.Unlike te clear . PreviouNUDT15*6 . The preNUDT15*3 , consistNUDT15*3 polymorphism had the increased risk of thiopurine-induced leukopenia and neutropenia in particular, as early as the first 2 months of the maintenance phase of treatment.The results from this meta-analysis showed that the lower number of OR was found in late onset of myelosuppression and that these may be because thiopurine toxicity often occurs in the first few months of the maintenance phase. The reason may be due to the fact that practically, the thiopurine dose was gradually adjusted to the tolerated dose which resulted in no myelosuppressive effect. Therefore NUDT15*3 carriers and early leukopenia. We could not explore the cause of this heterogeneity. Despite the fact that we used a random-effects model to pool the results across studies, the results of the meta-analysis for early leukopenia should be interpreted with caution.One limitation that deserves discussion is a significant heterogeneity observed amongst studies evaluating the association between with NUDT15*3 and NUDT15*2 variants and thiopurine-induced early onset myelosuppression. The rs116855232 SNP which exists in both of these variants appears to be a key SNP for thiopurine-induced hematoxtoxicity. The genotyping of the rs116855232 which has a high prevalence in Asian populations should be considered prior prescribing thiopurine drugs in order to predict the risk of early myelosuppression.In summary, this study found a strong relationship between


D) gene polymorphisms and IgA nephropathy (IgAN) was conflicting. This pooled analysis was performed to explore this issue.  The connection between angiotensin-converting enzyme insertion/deletion . In Asian and Chinese patients, ACE I/D gene polymorphism was also correlated with IgAN vulnerability. Moreover, ACE D allele, DD, and II genotype were correlated with the progression of IgAN . Conversely, in Caucasian subjects, there was no link between ACE I/D gene polymorphism and the risk of IgAN.  27 studies with 2538 IgAN cases and 3592 controls were included. In overall subjects, ACE D gene polymorphism was correlated with the vulnerability and progression of IgAN in Asian and Chinese patients, and ACE D allele and DD homozygote genotype could be adverse factors for IgAN, while the II homozygote genotype could be an advantage factor. But, no significant association was found between ACE I/D gene polymorphism and IgAN in Caucasians. ACE I/ IgA nephropathy (IgAN) is a common type of glomerulonephritis globally; in the Pacific Asian region, the prevalence rate of IgAN is even more high . It is aD) polymorphism has been found, and based on this I/D polymorphic marker locus, three genotypes could be defined: DD, II homozygotes, and ID heterozygote [D gene polymorphisms, and these subjects with the D allele always have higher level of ACE activity [D polymorphism would affect the vulnerability of IgAN [D polymorphism in the occurrence of IgAN is still inconsistent and controversial. So, in this study, we included much more high-quality studies to further assess the influence of ACE I/D gene polymorphism on IgAN susceptibility.Angiotensin-converting enzyme (ACE) was a key enzyme that converted angiotensin I to angiotensin II. The ACE gene consisted of 26 exons and located on chromosome 17q23 . Single- of IgAN –10. HoweD, angiotensin-converting enzyme, insertion/deletion, gene, and gene polymorphism. The inclusion criteria of this study included the following: (1) the study included two comparison groups (IgA patients vs. control subjects or progression IgA patients vs. nonprogression IgA patients). (2) The number of ACE I/D genotypes could to be calculated. (3) The genotype distributions in control group was in accordance with Hardy–Weinberg Equilibrium (HWE).We searched relevant studies from various electronic databases  for eligible trials . Multiple search keywords were used: IgA nephropathy, renal disease, kidney, nephropathy, ACE, ACE I/D gene polymorphism on IgAN susceptibility, and four different genetic models were used: model 1, allele D vs. allele I; model 2, genotype DD vs. genotype ID + II; model 3, genotype II vs. genotype DD + ID; and model 4, genotype ID vs. genotype DD + II. When a P value was less than 0.05, it was considered as statistically significant for the pooled OR. The heterogeneity among various included studies was evaluated using a Q test. Additionally, Begg' s test was completed to evaluate the publication bias; when the P value of Begg' s test was less than 0.05, potential publication bias was most likely to exist [In this study, we used the Stata software to perform this pooled analysis. An odds ratio (OR) was calculated to evaluate the influence of ACE I/to exist .As shown in D gene polymorphism on IgAN risk was reported in 21 trials [D gene polymorphism was associated with IgAN risk in the general populations . Additionally, the funnel plot was asymmetrical  polymorphism (rs4340) of a 287 bp Alu in the 16th intron was the most investigated [D polymorphism, and in those patients with the ACE DD genotype, the level of ACE expression and activity both in renal tissue and in serum were all markedly increased [D allele carriers had higher ACE level and consequently possessed higher risk to IgAN vulnerability.Recently, the researchers found that IgAN was also influenced by a confluence of environmental and genetic factors. Familial aggregation of IgAN indicated that genetic factor might participate in the pathogenesis of IgN . With thncreased ,45. The D polymorphism of ACE gene locus and IgAN risk was inconsistent in different races.In this study, the pooled analysis results indicated that the relationship between I/D were associated with the vulnerability and progression of IgAN, but there was no evident correlation in Caucasian patients. This might be due to the differences in the environment and genetic backgrounds. Also, ACE inhibitors could affect the analysis results, while some studies in Chinese subjects did not clearly state whether to exclude those patients using ACE inhibitors. In short, the understanding about the ACE I/D genetic susceptibility difference was acquainted scarcely, and future research is needed to explore the specific mechanisms.In Asian and Chinese patients, we found that the genotype DD, II, and allele D genetic alteration, and the relationship between many other genes SNPs and IgAN susceptibility needed to be further explored.Some potential limitations should be discussed in our study. Firstly, the sample sizes in some included trials were small. Secondly, public bias was existed in this meta-analysis. Third, the included trials were from various countries, but the trials were published only in English and Chinese, which would result in reporting bias. Fourth, some included studies did not clearly state whether to exclude those subjects using ACE inhibitors. Finally, our study was mainly focused on ACE I/D polymorphisms were linked with the susceptibility and progression of IgAN in Asian and Chinese individuals. ACE gene locus allele D and genotype DD could be risk factors of IgAN vulnerability. Conversely, those subjects carrying the ACE II genotype would reduce the risk of IgAN susceptibility. However, no obviously correlation was found between ACE I/D gene polymorphism and IgAN in Caucasian patients. In future studies, much more high-quality trials are required to clarify it.Our pooled analysis supported that ACE gene locus I/


Q and I2 statistic. We utilized the random-effect model to calculate the overall RR with 95% CI.  We systematically searched articles on electronic databases such as PubMed, Embase, Scopus, and Google Scholar between January 1, 2000 and July 30, 2020. Articles were independently evaluated by two authors. We included observational studies (case-control and cohort) and calculated the risk ratios (RRs) for associated with anemia and PD. Heterogeneity among the studies was assessed using the N = 4, RRadjusted = 1.17 . However, heterogeneity among the studies was significant . The pooled relative risk of PD in female patients with anemia was higher  as compared to male patients with anemia .  A total of 342 articles were identified in the initial searches, and 7 full-text articles were evaluated for eligibility. Three articles were further excluded for prespecified reasons including insufficient data and duplications, and 4 articles were included in our systematic review and meta-analysis. A random effect model meta-analysis of all 4 studies showed no increased risk of PD in patients with anemia ( This is the first meta-analysis that shows that anemia is associated with higher risk of PD when compared with patients without anemia. However, more studies are warranted to evaluate the risk of PD among patients with anemia. Parkinson disease (PD) is the second most common age-related neurodegenerative disorder, with more than 60,000 newly diagnosed cases yearly in the USA . The incThe common risk factors of developing PD are exposure to pesticides, genetic variants, environment toxins, idiopathic REM sleep behavior disorder (RBD), and focal cerebrovascular damage –6. HowevAnemia hampers erythropoiesis and enhances eryptosis, whereas insufficient iron is reported in the substantia nigra of PD patients , 17. IncGoal: The aim of this current systematic review and meta-analysis is to evaluate the published epidemiological studies for clarifying the association between anemia and risk of PD.Study the magnitude of the risk of PD in patients with anemia and without anemiaCalculate the magnitude difference of the risk of PD in male and female patients with anemiaReduce the confounding factors by evaluating the risk based on various adjustmentsWe did a systematic search on electronic databases such as PubMed, Embase, Scopus, and Web of Science between January 1, 2000 and July 30, 2020. The following search terms were used to collect relevant articles: “anemia,” OR “low level hemoglobin,” “iron deficiency” And “Parkinson disease.” The initial search was conducted by one author who is an expert in systematic review. Moreover, we searched in the reference lists of retrieved articles to ensure the comprehensiveness.The eligibility criteria were restricted to observational studies (case-control and cohort) and clinical trials  that evaluated the association between anemia and PD as a primary outcome. Studies were included if they were published in the form of (a) original study, (b) participants at least 200, (c) published in English, (d) provided clear definition of anemia and PD, and (e) provide proper effect size to summarized pool risk.We excluded studies if they published in the form of review, short report, poster, editorial, case-report, and correspondence.Two authors (TNP and YCW) independently reviewed all the titles and abstracts of retrieved articles. They used prespecified selection criteria to select relevant articles to be included in this meta-analysis. Any disagreement in this stage was resolved by the difference by discussing with third author.Same two authors developed data collection form for extracting the required data from the selected studies. They checked the study duplication by comparing authors' names, publication year, and location of study. Two authors collected the information about effect size , adjusted factors, total number of participants, number of anemia and PD patients, inclusion and exclusion criteria of anemia and PD, age, percentage of male, duration of study period, and location.Q and I2 statistic. Finally, publication bias evaluated by 3 funnel plot–based methods: the Egger's test, the Begg's test, and the trim and fill method.The Newcastle-Ottawa Scale (NOS) was used to assess the quality of each study. They calculated the NOS score and divided into two groups: low and high qualities. The heterogeneity was calculated among study-specific RRs using the Q statistic and quantified by the I2 value. The heterogeneity among the studies was categorized into four groups, namely, very low (<25%), low (25~50%), medium (50~75%), and high (>75%) [p value less than 0.5 is considered as a significant.The meta-analysis was performed on selected studies that evaluated the magnitude of the association by using adjusted ORs and HRs. The pooled risk ratio (RR) was calculated from selected studies to show the effect size. The random-effect model was used in this meta-analysis. We calculated statistical heterogeneity across the various studies which were tested using the Cochran h (>75%) –25. WhenInitial search in the electronic databases yielded 342 articles. A total of 335 articles were excluded after reviewing the titles and abstracts of retrieved articles. However, 7 articles went to full-text review, and 3 articles were further excluded due to not matching the prespecified selection criteria. Finally, 4 articles were included in the current systematic review and meta-analysis –29.  FigThis current systematic review and meta-analysis comprised of 4 studies including 3 cohorts and 1 case-control study . IncludeN = 4, RRadjusted = 1.17 . However, heterogeneity among the studies was significant   .N = 3, RRadjusted = 1.14  as compared to male patients with anemia   and female  patients with anemia.We pooled 3 studies in order to evaluate the risk of PD based on gender. The pooled relative risk of PD in female patients with anemia was higher ( = 0.51) . HoweverN = 3, RRadjusted with diabetes and hypertension = 1.10 , and heterogeneity was significant .We also evaluated the risk of PD in patients among the studies adjusted with most common confounding factors such as diabetes and hypertension. The pooled risk ratio was  all included studies quality were high and (b) consider adjusted effect size; therefore, risk of bias is low and (c) clear definition of anemia and PD risk.α-synuclein and etiology of PD are the main keys that can be used to explain the pathological mechanism of anemia and PD risk. Erythrocytes generate maximum portion of α-synuclein in the human blood [α-synuclein oligomer and total protein is higher in erythrocyte of PD patients when compared to normal patients [α-synuclein in PD has received wide attention over the past years. A previous mice study showed that microbial metabolites may increase the risk of neuroinflammation by expressing proinflammatory cytokines which ultimately lead to the development of motor symptom [The biological mechanism of PD and anemia is not fully understood yet. However, there are several biological explanations that can define their association. First, the neural degeneration and disease progression among the PD patients can be induced by either apoptosis or necrosis . The braan blood ; it has patients . Anxietypatients , 41. Deppatients . Moreove symptom . Other s symptom .PD is now one of the leading causes of disability and deaths globally. A significant amount of epidemiological studies have highlighted an increased prevalence of PD which raised sheer concern and urgent need for public health strategies –47. ApprOur first and rigorous meta-analysis has several strengths. First, this is the first meta-analysis that investigated the association between anemia and PD risk. Second, this study included four large observation studies from Taiwan, Israel, South Korea, and USA; healthcare data quality of these countries is world standard. They adjusted potential confounding factors to this study to reduce bias of effect size calculation. Third, this present study showed risk difference between male and female patients.However, this study has some limitations. First, number of included studies is only four, although all are high quality study design with larger number of participants. Second, we were not able to provide risk of PD based on severity of anemia  and BMI . Third, the pooled risk was not calculated based on location, study design, and other factors like smoking status, alcohol consumptions, and duration of anemia; it is because data were limited.Our study shows that the risk of PD was higher among the patients with anemia as compared without anemia. The risk of PD was higher in male patients. Therefore, more protective strategies should be taken in male patients when compared to female patients. The risk of PD among patients with anemia can be explained by some biological mechanisms like oxidative stress and downregulation of iron homeostasis. More observational studies in different regions and biological studies are warranted to clarify the mechanism underlying their association.


Background: The atherogenic index of plasma (AIP), which is the logarithm of the ratio between the triglyceride and high-density lipoprotein cholesterol (TG/HDL-C) concentrations in molar units, is correlated with the burden of atherosclerosis. This study aimed to evaluate the association between the AIP and coronary artery disease (CAD) in the adult population by performing a meta-analysis.Methods: Observational studies relevant for this meta-analysis were identified by searching the PubMed, Embase, and Web of Science databases. Only studies using multivariate analysis were considered. A random-effects model, which incorporates potential intra-study heterogeneity, was applied to combine the results.Results: Ten observational studies were included. In studies with the AIP analyzed as a continuous variable, a higher AIP was associated with a higher odds of CAD . Further analysis of studies with the AIP analyzed as a categorical variable showed a higher odds of CAD  in the participants with the highest versus the lowest AIP value. Subgroup analyses demonstrated consistent results in asymptomatic and symptomatic populations as well as in male and female participants .Discussion: Current evidence, mostly from cross-sectional studies, suggests that a higher AIP value may be independently associated with CAD in the adult population. Coronary artery disease (CAD) remains one of the major causes of morbidity and mortality in the global population –3. The bThe Meta-analysis of Observational Studies in Epidemiology  StatemenPubMed, Embase, and the Web of Science databases were searched using a combination of the following search terms: (1) “atherogenic index plasma” OR “AIP”; and (2) “atherosclerotic cardiovascular disease” OR “ASCVD” OR “cardiovascular events” OR “MACE” OR “cardiovascular” OR “coronary artery disease” OR “CAD” OR “CHD” on May 10, 2021. Only studies published in English were included. The references of related literature reports were also examined manually for possible available studies for the meta-analysis.The inclusion criteria were as follows: (1) design and form: observational studies in full-length articles; (2) participants: adults in a community or a hospital setting; (3) exposure: AIP; (4) outcome: CAD; and (5) data reported: the association between the AIP and CAD in multivariate analysis. The diagnosis of CAD was consistent with that applied in the original studies.Two authors performed the database search, literature review, data input, and study quality scoring independently, in accordance with the predefined criteria. If a disagreement occurred, it was resolved by consensus between the two authors. The following data were obtained for each study: (1) general information of the study; (2) characteristics of the study design; (3) participant information, such as the number of adults included, health status, and demographic features; (4) AIP presentation form ; (5) definition of CAD; and (6) variables adjusted in the multivariate analyses. The Newcastle–Ottawa Scale (NOS)  was appl2 statistic  and the 95% confidence interval (CI) were used to indicate the association between the AIP and CAD. If the AIP was analyzed as a continuous variable in the original studies, the RR of CAD per 1-standard deviation (SD) increment of the AIP was used. If the AIP was analyzed as a categorical variable in the original studies, the RR of CAD in participants with the highest versus the lowest AIP level was extracted. Between-study heterogeneity was assessed by Cochrane's Q test and estimation of the Itatistic . Signifi RR data . By omit RR data . Subgrou RR data . RevMan The flowchart of the literature search and selection is shown in P < 0.001; 2 = 90%). Sensitivity analysis by excluding one dataset at a time showed consistent results . In addition, subgroup analyses showed consistent results in asymptomatic and symptomatic populations as well as in male and female participants . Sensitivity analysis by omitting one dataset at a time did not significantly affect the results . Furthermore, subgroup analyses did not show that the health status or the sex of the participants significantly influenced the association between the AIP and CAD .As shown in In this meta-analysis of observational studies, we found that a higher AIP was independently correlated with higher odds of CAD after adjusting for conventional cardiovascular risk factors. The results with AIP as a continuous variable compared to as a categorical variable were consistent. Further sensitivity analyses by excluding one study at a time did not significantly affect the results, indicating the robustness of the findings. In addition, the results of predefined subgroup analysis showed that the association between the AIP and CAD was not significantly affected by the health status or the sex of the patients. Taken together, these results support that the AIP may be an independent risk factor for CAD in the adult population.To the best of our knowledge, this may be the first meta-analysis that summarizes the current knowledge regarding the association between the AIP and CAD in the adult population. Several methodological strengths of this study should be noted before the clinical implications of the findings are discussed. First, an extensive search strategy was applied, aiming to incorporate all eligible studies. In addition, studies with the AIP analyzed as a continuous variable and as a categorical variable were pooled separately, and the consistent results further indicated the validity of the findings. Moreover, only studies with multivariate analysis were included, thus providing a potentially independent association between the AIP and CAD. Finally, multiple sensitivity and subgroup analyses further indicated the stability of the findings. The results of the meta-analysis were consistent with previous studies showing a close correlation between the AIP and cardiovascular risks. For instance, a cross-sectional study including 340 healthy women has revealed that the AIP level is significantly correlated with the Framingham risk score (FRS), indicating the possible role of the AIP in the early diagnosis of CAD . The assOur study also has some limitations that must be addressed. First, only one of the included studies was a cohort study. A sequential relationship between a higher AIP value and an increased risk of CAD should be further validated in large-scale prospective studies. Second, the meta-analysis was based on data at the study level rather than for individual patients. We could not determine whether a difference in participant characteristics, such as age, ethnicity, and comorbidities, could be a source of heterogeneity among the included studies. In addition, although only studies with multivariate analysis were included, there might be residual factors that may confound the association between the AIP and CAD, such as dietary factors and concurrent medications that may affect TG and/or HDL-C levels. Finally, an optimal cutoff value of the AIP for the prediction of CAD risk remains to be determined, and it continues to be unknown whether incorporation of the AIP into the current risk stratification process for CAD, such as the FRS, is associated with an improved predictive efficacy. Therefore, future studies are necessary.In conclusion, the results of this meta-analysis suggest that a higher AIP value may be independently associated with a higher odds of CAD in the adult population. Large-scale prospective cohort studies are warranted to validate these findings and to evaluate the possible role of the AIP as a predictor for CAD risk in the general adult population.The original contributions presented in the study are included in the article/JWu, QZ, ZW, and MC conceived and designed research. JWu, QZ, and JWe collected data and conducted research. JWu, ZW, and MC analyzed and interpreted data. JWu and QZ wrote the initial paper. MC and QZ revised the paper. MC had primary responsibility for final content. All authors read and approved the final manuscript.This work was supported by China National Natural Science Foundation (81702597) and Research Fund of the First Hospital of Jilin University (04020940001 and JDYY11202036).The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.All claims expressed in this article are solely those of the authors and do not necessarily represent those of their affiliated organizations, or those of the publisher, the editors and the reviewers. Any product that may be evaluated in this article, or claim that may be made by its manufacturer, is not guaranteed or endorsed by the publisher.


To perform a systematic review and network meta-analysis to characterize the effect of novel androgen receptor axis-target (ARAT) agents on diarrhea and constipation.ClinicalTrials.gov up to September 2021 for phase 3 randomized controlled trials (RCTs) of patients receiving novel ARAT agents for prostate cancer (CaP). A Cochrane risk-of-bias tool was used to assess trial quality. The primary outcomes were risk ratio (RR) of any-grade diarrhea and constipation for patients receiving ARAT treatment. RRs of competing treatments were evaluated by pairwise and Bayesian network meta-analysis.We searched the Pubmed, Web of Science, and RR = 1.30, 95% CI ). As for subgroup analysis, abiraterone, enzalutamide, and apalutamide were all associated with significant increased risk of any-grade diarrhea . Based on Bayesian modeling, abiraterone and enzalutamide showed the highest and lowest probability to rank first in terms of increasing risk of any-grade diarrhea. There were no significant differences of risk in any-grade constipation, grade 3 or greater diarrhea, and constipation between ARAT and control group.In this study, 13 trials with 15,117 participants comparing 5 treatments  were identified. Use of novel ARAT agents was associated with a significant increased risk of any-grade diarrhea  and castration-resistant prostate cancer (CRPC). Prostate cancer (CaP) is the most common cancer in men, which accounts for 26% of diagnoses in the United States in 2021 . In receImproved prognosis has created growing needs to address the unique health issues facing CaP survivors that result from CaP, its treatment, and related comorbid conditions. Previous studies have demonstrated that gastrointestinal (GI) complications, such as diarrhea and constipation, are one of persistent burdens for CaP survivors treated with ADT –11. AlthThis study aimed to determine the effect of novel ARAT agents on GI complications utilizing the reconstructed clinical data derived from phase 3 RCTs to inform decision-making. As most of included trials only report part of GI complications, we focus on two of the most reported complications which are diarrhea and constipation. Additionally, diarrhea and constipation are both most concerned GI complications of cancer therapies.ClinicalTrials.gov up to September 2021. The following searching terms were used: [“Prostate Cancer”] AND [“Abiraterone” OR “Enzalutamide” OR “Apalutamide” OR “Darolutamide”]. We performed the study eligibility using the population, intervention, comparator, outcome, and study (PICOS) approach: (P) studies focused on patients with a diagnosis of CaP; (I) treated with Abi, Enz, Apa, or Dar; (C) in which placebo was performed as a comparator; (O) reporting one or both of the following outcomes: diarrhea and constipation; (S) in phase 3, placebo-controlled, double-blind, and randomized trials.We conducted a systematic review and network meta-analysis followed the Preferred Reporting Items for Systematic Reviews and Meta-analyses (PRISMA) reporting guideline and its extension for network meta-analysis , 20. We Two investigators independently conducted title and abstract selection and full-text review. The PRISMA flowchart about the selection process are displayed in RR) for any-grade diarrhea and constipation of patients receiving any types of novel ARAT agents compared with control group. The secondary outcomes included: (a) combined incidence of any-grade and grade 3 or greater diarrhea or constipation in the total ARAT group, ARAT subgroups , and control groups; (b) RR for grade 3 or greater diarrhea and constipation of patients receiving any types of novel ARAT agents. Any-grade adverse events are defined as events from grade 1 to higher grades.The primary outcomes of the present meta-analysis were risk ratio (Risk of bias (RoB) was independently determined by two investigators using the Cochrane Collaboration's tool (version 2.0) . To asseCIs were estimated for diarrhea and constipation using pairwise and network meta-analysis. The analysis was performed in two steps: first, pairwise meta-analysis was performed to assess a particular outcome. The Mantel–Haenszel (M–H) random-effects model was applied to pooled RRs for any-grade and grade 3 or greater diarrhea or constipation, given the expected heterogeneity within part of the evaluated trials. Heterogeneity across studies was formally tested for using chi-square (p < 0.05) and the I2 statistic. Subgroup analysis was based on the different types of ARAT agents that patients received in intervention groups. Publication bias was evaluated utilizing the funnel plots and the Peters test of funnel plot symmetry  and Review Manager v5.2 software.The significant level was A total of 702 unique records were screened for eligibility, of which 651 were excluded based on the evaluation of titles and abstracts. Full-text screening was assessed for 51 articles, eventually, 15 unique studies (such as 13 RCTs) fulfilled inclusion criteria .In total, 13 trials comparing five treatments were assessed, including placebo, Abi, Enz, Apa, and Dar . The welp were 0.82 and 0.63, respectively) (ctively) , 2.CI [23.7–27.2%]), followed by Apa at 20.3% , followed by Enz at 14.5% , followed by Dar at 7.4%  , followed by Enz at 17.5% , Apa at 10.1% , and Dar at 6.9%   and that of grade 3 or greater diarrhea was 45 (0.5%). In the control group, the reported any-grade diarrhea was 756 (14.5%) and that of grade 3 or greater diarrhea was 24 (0.3%). The reported incidence of constipation in the ARAT group was 1,463 (16.9%) and that grade 3 or greater constipation was 28 (0.2%). The reported incidence of constipation in the control group was 991 (15.3%) and that grade 3 or greater constipation was 24 (0.2%) . As for , 9.3%]) . Similar, 9.3%]) . As for , 8.7%]) . The seq, 8.7%]) .RR = 1.30, 95% CI , I2 = 32%)   (p = 0.321)  . As for , 2.00]) . As for I2 = 0%) . Subgrou= 0.321) .RR = 1.55, 95% CI )  . Althoug, 2.13]) . Further, 2.13]) . Based o, 2.13]) . Last, wp = 0.992)  .Network meta-analysis showed consistent results with pairwise meta-analysis between ARAT and control group . SimilarAcross 13 RCTs, such as patients with CaP treated with novel ARAT agents, we found that the use of ARAT agents was associated with the risk of diarrhea. The use of ARAT agents was associated with 1.30 times higher risk of diarrhea. As for individual ARAT agents, Abi, Enz, Apa, and Dar were associated with 1.40, 1.17, 1.35, and 1.33 times higher risk of diarrhea, respectively. There were no significant differences for risk of constipation between ARAT and control groups. Based on Bayesian modeling, Abi might be associated with highest risk of diarrhea, among the four ARAT agents.RR of each ARAT agents for risk of diarrhea. The present study highlighted several insights. First, we evidenced that novel ARAT agents could significantly increase the risk of diarrhea for patients with CaP based on data from phase 3, placebo-controlled, double-blind, and randomized trials. There were potential links between CaP and treatment and increased inflammatory levels from GI dysbiosis . Fourth, there were no time-based data to calculate the diarrhea and constipation person-year incidence rates. Fifth, the present study only included 2 and 1 trials focused on Apa and Dar, therefore, it might need further update. Sixth, the present results are from network meta-analysis, therefore, prospective clinical trials regarding this issue are suggested in the future.Our study has some limitations. First, clinical consequences of diarrhea and constipation on therapy, and the use of potential interventions were not reported in our included studies. Therefore, the two potential interventions that we mentioned needs further investigations. Second, we could not conduct age-stratified or other subgroup analysis for risk of diarrhea or constipation, because the cut-off levels were different across trials and the included studies were not focused on reporting risk factors for diarrhea and constipation related to age or other valuables. Third, the duration of treatment was different across trials. The relatively short therapy duration for novel ARAT agents may bias against their long-term effectiveness estimation. Interestingly, meta-regression analysis regarding the duration of hormone therapy both for risk of diarrhea and constipation indicated that the duration of hormone therapy might not affect the stability of our present results , the National Natural Science Foundation of China , the Post-doctoral Science Research Foundation of Sichuan University (2020SCU12041), the National Clinical Research Center for Geriatrics, West China Hospital, Sichuan University (Z2018C01), the Technology Innovation Research and Development Project, Chengdu Science and Technology Bureau (2019-YF05-00296-SN), the Clinical Research Incubation Project of West China Hospital of Sichuan University (2020HXFH033), and the China Society of Clinical Oncology—Pilot Oncology Research Fund (Y-2019AZMS-0523).The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.All claims expressed in this article are solely those of the authors and do not necessarily represent those of their affiliated organizations, or those of the publisher, the editors and the reviewers. Any product that may be evaluated in this article, or claim that may be made by its manufacturer, is not guaranteed or endorsed by the publisher.