Systematic Review Registration:https://www.crd.york.ac.uk/prospero/display_record.php?ID=CRD42021228391, identifier CRD42021228391.The use of prophylactic cranial irradiation (PCI) for small cell lung cancer (SCLC) patients is controversial. Risk factors for brain metastasis (BM) development are largely lacking, hampering personalized treatment strategies. This study aimed to identify the possible risk factors for BM in SCLC.We systematically searched the Pubmed database (1 January 1995 to 18 January 2021) according to the PRISMA guidelines. Eligibility criteria: studies reporting detailed BM data with an adequate sample size  in patients with SCLC. We summarized the reported risk factors and performed meta-analysis to estimate the pooled hazard ratios (HR) if enough qualified data  were available. In total, 61/536 records were eligible , in which 57 factors were reported. Ten factors qualified BM data for meta-analysis: Limited stage disease (LD)  and older age (≥65)  were associated with less BM; A higher T stage (≥T3)  was a significant risk factor for BM. Male sex  tended to be a risk factor, and better PS (0–1)  tended to have less BM. Smoking, thoracic radiotherapy dose were not significant (P >0.05). PCI significantly decreased BM (P <0.001), but did not improve OS in ED-SCLC (P = 0.81). A higher PCI dose did not improve OS (P = 0.11). The impact on BM was conflicting between Cox regression data  and competing risk regression data . Compared to M0–M1a, M1b was a risk factor for OS (P = 0.01) in ED-SCLC, but not for BM (P = 0.19). As regular brain imaging is rarely performed, high-quality data is lacking. Other factors such as N-stage and blood biomarkers had no qualified data to perform meta-analysis. In conclusion, younger age, higher T stage, and ED are risk factors for BM, suggesting that PCI should be especially discussed in such cases. Individual patient data (IPD) meta-analysis and well-designed RCTs are needed to better identify more risk factors and further confirm our findings.  Small cell lung cancer (SCLC) accounts for about 13% of newly diagnosed lung cancers worldwide . PatientWe conducted this study according to the PRISMA guideline   and regiWe extracted data according to our published protocol  and repoTwo investigators (HZ and DZ) independently screened the titles, abstracts, methods, and full texts for eligibility; extracted data; and assessed the risk of bias. Any conflicts in each step were resolved through discussion with a third investigator (LH).2 to quantify inter-study heterogeneity, of which 25, 50, and 75% can be considered low, moderate, and high heterogeneity  –27. As sIn addition to symptomatic BM, we found that the pre-PCI BM (BM immediately before PCI) was investigated in one study  and the We also found that the definition of time to BM events varied among studies, which indicates that heterogeneity also exists between RCTs: from the date of initial diagnosis n = 19) , 78, 81; , 78, 812-test or Fisher exact 2-tailed test: n = 7  , 59–61. 2. Sex: Sex was investigated in 16 studies. It concluded that sex was not an independent risk factor for BM or OS in 13 studies , 59–61. 3. Smoking: Smoking was investigated in seven studies. It has been shown that smoking is not a significant risk factor for BM or OS , 56, 61.1. TNM cT stage: The T stage was investigated in four studies with conflicting conclusions , 52, 55.2. c-stage: c-stage was investigated in different ways in 11 studies with conflicting conclusions , 64, 65 3. M-status in ED-SCLC: M status (M1b or M0–M1a) was investigated in patients with ED-SCLC in four studies , 61, 48.4. PS: PS was investigated in 10 studies in different ways. It was concluded that PS was not a significant risk factor for BM or OS in six SCLC studies , 55, 63.1. PCI vs no PCI: PCI was investigated in 28 studies, including 8 RCTs. Three RCTs had qualified overall BM data for meta-analysis based on Cox regression , 34, 68 2. PCI dose: PCI dose was investigated in four RCTs , 34, 68 3. TRT dose: TRT dose (<45 Gy vs ≥45 Gy) was investigated in patients with SCLC in two studies , 52 and The other 47 factors did not have sufficient qualified data to perform meta-analysis, such as N-stage, number of distant metastasis, and blood biomarkers. Detailed reasons are summarized in Data on risk factors for BM in SCLC are largely lacking, which makes personalized treatment  difficult. It also impairs the design and interpretation of RCTs evaluating PCI. We identified several factors that were associated with a higher risk of BM: higher T-stage, ED, male sex, and younger age. As has already been reported previously , 82, we per se, resulting in a short OS, making M-status factors less relevant than risk factors for BM development.It is not surprising that ED and higher T stage, which means more advanced tumor load, were risk factors for BM. It is more interesting to note that compared to M0–M1a, M1b was a risk factor for OS but not for BM in patients with ED-SCLC. This could be explained by the aggressive nature of ED-SCLC We also found younger age (<65) as a risk factor for BM. This is probably because younger SCLC patients generally live longer , 58 and Similarly, the cut-off value of PS also varied among studies, resulting in only PS ≥2 having qualified data to perform meta-analysis based on two retrospective studies. It showed that worse PS (≥2) tended to be at a higher risk of BM. This is at odds with a secondary analysis of the CONVERT trial showing that poorer PS (1–2 vs 0) patients had a lower risk  of brain progression , likely We also showed a marginally significant risk of developing BM in males. This is consistent with former reports illustrating that female patients had better prognosis than males, in SCLC , NSCLC , or otheFurthermore, we found that PCI reduced BM in SCLC but did not improve OS in ED-SCLC, which is based on the EORTC phase III trial  and the Interestingly, we noticed that the meta-analysis results based on competing risk regression and Cox regression could be different, which indicates that data based on different statistical analysis methods should not be pooled together to perform meta-analysis. In this current study, only PCI dose (≤25 Gy vs >25 Gy) had qualified data to perform meta-analysis for both regressions. The Cox regression data showed that PCI dose was not a significant risk factor for BM , while the competing risk regression data showed that a higher dose (>25 Gy) could prevent BM more effectively . Of note, both analyses contained the same RCT conducted by Le Pechoux et al. , in whicPCI best timing is also unknown. Current guidelines do not have a definite consensus on this issue . We idenSimilarly, four RCTs –33, 35 aRisk of bias assessment is essential in systematic reviews and meta-analyses. We assessed the risk of bias for RCTs using the RoB2 tool and noticed that it has its limitations. It assesses the process of data collection and data reporting but does not assess the methods of data analysis. However, inappropriate analysis can lead to different/misleading conclusions. It also does not evaluate trials that were closed earlier, which results in much less powerful conclusions. Therefore, the improvement of the RoB2 tool is needed to assess the risk of bias more thoroughly and help improve the design of RCTs.As for the non-RCTs, Wells et al. proposed the Newcastle–Ottawa-Scale (NOS) for assessing the quality on a website rather than in a peer-reviewed journal . Till noThe Cochrane community recommends the Risk Of Bias In Non-randomized Studies of Interventions (ROBINS-I) tool for assessing the risk of bias in non-RCTs of interventions . HoweverAdditionally, current risk of bias assessment tools mainly assesses the risk of bias per study. This is fine for studies that mainly investigate interventions. However, as a meta-analysis aims to identify all related risk factors, it is necessary to assess the risk of bias per factor in each study. Therefore, we assessed the quality of data per factor, mainly focusing on the analysis methods in each study and summarized the possible problems in the comments. In this way, readers can clearly interpret the results.As far as we are aware, this is the first systematic review and meta-analysis to identify risk factors for BM in SCLC. Most current meta-analyses focused on one aspect, such as PCI or not in SCLC , ED-SCLCAdditionally, we first used a simple and effective method to assess the quality of data before pooling everything together to perform the meta-analysis. That is, only studies of the same type using the same method with proper statistical analysis should be pooled together under the premise that the patients belong to the same category. This will avoid misleading conclusions based on heterogeneous data.Furthermore, we noticed that many studies retrieved in our search  did not report BM-related outcomes. Moreover, brain imaging is often lacking in published studies. To evaluate BM risk factors better, it is very crucial to document baseline characteristics, treatment, as well as adequate and regular brain imaging. Brain imaging should be preferred over MRI, as this is the best imaging modality to detect asymptomatic BM. Regular brain imaging is important in clinical trials, as even after a negative baseline brain MRI, in a study by Manapov et al., the second cranial MRI after completion of chemoradiotherapy revealed asymptomatic BM in 11/40 (32.5%) LD-SCLC complete responders . In someWe also noticed that many studies which reported BM data did not report OS data. This hampers the interpretation of clinical significance. For example, if a factor (A) is a risk of BM but not for OS, a factor (B) is a risk of both BM and OS, and another factor (C) is a risk of BM but unknown for OS, clinicians will put much higher weight on considering factor B and much less weight on considering C when making an individualized management strategy. Therefore, we suggest researchers report OS data as well when reporting BM data to enhance the clinical application value.In conclusion, multiple studies evaluated risk factors for SCLC BM, but limited data were qualified to perform a meta-analysis. We found that younger age, higher T stage, and ED were risk factors for BM; suggesting that PCI should be especially discussed in such cases, shared decision making is necessary; and that higher PCI dose is not necessary. IPD meta-analysis and well-designed RCTs with high-quality data are needed to identify more risk factors such as blood biomarkers, and confirm our findings. Regular MRI with contrast-enhancement before PCI and during follow-up is helpful to detect asymptomatic BM, especially for patients with a high risk for BM. The MRI compliance at each pre-specified time point should also be reported in prospective trials. Better collaboration with statisticians is needed in future studies. We suggest emendation of the ROB2 tool to assess the statistical methods as well.HZ, DDR, and LH conceived this study. HZ and DDR searched papers in Pubmed. HZ and DZ screening the papers from titles to full texts, extracted the data, and assessed the risk of bias. LH checked the screening, extraction and assessments. HZ, WW, and RH analyzed the results. DDR and LH supervised the whole process. HZ, LH, and DDR draft the manuscript. AL, AT, WW, RH, FMK, and DZ made the revisions. All authors listed have made a substantial, direct, and intellectual contribution to the work and approved it for publication.This research was supported by the following grant: Scholarship of China Scholarship Council (Grant No.: CSC 201909370087).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.


Patients with gallstone disease (GSD) often have highly co-occurrence with metabolic syndrome (MetS) and Nonalcoholic fatty liver disease (NAFLD) both associated with insulin resistance (IR). Meanwhile, highly prevalence of NAFLD was found in patients who received cholecystectomy. However, the associations of GSD with MetS, NAFLD is inconsistent in the published literature. And risk of cholecystectomy on NAFLD is unclear.We searched the Medline EMBASE and WOS databases for literature that met our study topic. To be specific, studies with focus on associations between GSD and MetS/NAFLD, and risk evaluation on cholecystectomy and NAFLD incidence were enrolled for further analysis. The random effect model was used to calculate the combined relative ratio (RR) and odds ratio (OR)and 95% confidence interval (CI).2 = 41.1%, P=0.165). Risk of GSD was increased by 52% in NAFLD patients . And about 32% of increment on NAFLD prevalence was observed in patients with GSD . With regard to individual MetS components, patients with higher systolic blood pressure were more prone to develop GSD, with combined SMD of 0.29 . Dose-response analysis found the GSD incidence was significantly associated with increased body mass index (BMI)  in linear trends. Patients who received cholecystectomy had a higher risk of post-operative NAFLD , P<0.05). And this impact was amplified in obese patients .Seven and six papers with focus on connections between GSD and NAFLD/MetS prevalence. Correspondingly, seven papers with focus on risk of cholecystectomy on NAFLD occurrence were also enrolled into meta-analysis. After pooling the results from individual study, patients with GSD had higher risk of MetS (OR:1.45, 95%CI: 1.23-1.67, Ide-novo NAFLD after cholecystectomy.Our results confirmed that controls on weight and blood pressure might be candidate therapeutic strategy for GSD prevention. And concerns should be raised on  GSD is Clinically, insulin resistance (IR) is defined as the inability of insulin to keep blood glucose levels in a healthy range . HoweverConversely, systemic glucose and lipid metabolism can be regulated by gallbladder . The galIn view of the tight relationship between GSD and metabolic derangements, many studies were performed with topics on associations between GSD and MetS/NAFLD occurrence –36. Othe22.1A Meta-analysis was conducted according to the guidelines of the Preferred Reporting Items for Systematic Reviews and Meta-analyses (PRISMA)  . A relev2.2Eligibility criteria: 1) Published retrospective, prospective cohort studies and cross-sectional studies. 2) GSD / METS / NAFLD was the testing group's endpoint. 3) MetS diagnostic criteria for the study were given and the diagnosis of cholelithiasis needed to be confirmed by imaging or surgery. 4) The odds ratio (OR) / relative ratio (RR) / Hazard Ratio (HR) and corresponding 95% confidence intervals (CI) may be derived from studies or can be calculated. Studies were excluded if 1) literature was not the above research type, or the unpublished; 2) The subjects were not related to GSD/METS/NAFLD (no interest of subjects); 3) The disease outcomes were not just GSD/METS/NAFLD but with organic lesions, liver dysfunction or viral hepatitis in the hepatobiliary system (no interest of outcomes).The study characteristics were extracted from all the literatures: first author; publication year; country and ethnicity of data origin; study design type; enrolled study population ; disease outcome; definition of MetS and NAFLD; risk of disease; the mean ± SD of MetS components such as body mass index (BMI), blood pressure , waist circumference (WC), triglyceride (TG), fasting blood glucose (FBG), high density lipoprotein cholesterol (HDL-C) in case and control group; the calculation method of origin data; and adjusted covariates .2.3A preliminary assessment of the quality of each study was evaluated by two authors (JTL and QHL), respectively using the Newcastle-Ottawa Scale (NOS) . NOS con2.42 statistics , and chose OR  and RR (for prospective cohort study))and 95% confidence intervals to quantify the relationship between Gallstone and MetS, gallstone and NAFLD, and cholecystectomy and NAFLD. If heterogeneity is significant, the random-effects model will be used, and heterogeneity will be assessed by Iectively .2 increase, and then OR increments from both studies were combined to reflect the linear dose-response risk for GSD incidence. On the basis of information from reviewed studies, pooled standardized mean differences (SMD) of specific individual MetS components connected to the GSD occurrence were analyzed. In addition, a subgroup analysis was performed to determine the impact of potential confounding factors. Sensitivity analysis was carried out to look into the effects that a single study could have had on the outcomes.To explore the relationship between GSD and MetS/NAFLD, we first compared the combined OR of end-stage (MetS/NAFLD) with or without GSD and the combined OR of end-stage (GSD) with or without GSD. Secondly, according to the different BMI intervals provided in the two reports , 30, theWhen discussing the relationship between cholecystectomy and NAFLD, the combined RR and RR of NAFLD patients with or without cholecystectomy were also compared. Subsequently, subgroup analysis and meta-regression were performed to explore the heterogeneity of potential sources further.To calculate any potential publication bias, Egger’s test was employed. P<0.05 was deemed significance. All of the statistical analyses were performed using Stata 15.0 version software .Our study flow diagram was shown in 33.1The flow diagram of the meta-analysis registration study was shown in 3.2The study characteristics are presented in There are six articles focusing on the relationship between GSD and NAFLD, two articles were prospective cohort studies, and four were cross-sectional studies. All patients were diagnosed with GS and NAFLD using Ultrasonogram (USG) Diagnosis. The total number of participants was as high as 322,630, including 215,497 patients with GSD, whose prevalence ranged from 4.44% to 76.66%, and 5926 patients with NAFLD, whose prevalence ranged from 1.86% to 34.43%. East Asians, Hispanics and Caucasians participated in four, one and one studies.Six articles were included to explore the occurrence of NAFLD after cholecystectomy. A total of 261002 people participated, and the probability of suffering from NAFLD after the operation was as high as 60.97% Those who were East Asian or Hispanics engaged in just two and four studies respectively. Alcohol intake and physical activity were adjusted as covariates in three studies and only two study excluded patients with type 1 diabetes.3.33.3.12=41.1%, P=0.165; The pooled OR of incident MetS was 1.45 (95%CI: 1.23-1.67) with low heterogeneity , which is the same as that of Westerners and low sample size Table 2.In order to further explore the causes of overall heterogeneity, we carried out a sensitivity analysis. After omitting one study in turn and re-evaluating the summary OR of other studies, it is found that the heterogeneity is eliminated by excluding Chen’s study   without heterogeneity, P=0.624 , without heterogeneity, P=0.370 , with high heterogeneity  Figure 9Sensitivity analysis showed that the heterogeneity of cross-sectional study was affected by Kwak et al.’s study and Carmen et al. ‘s study could significantly change the RR and heterogeneity of the aggregate . And GSD was a risk factor of MetS with low heterogeneity. However, we still performed subgroup analyses and the results showed that the heterogeneity was strengthened by sample size, GSD incidence and ethnicity. So we supposed that the epidemiological evidence of MetS could make sense. Prevalence among white adults in developed countries was as high as 10% to 15% . And TheA liver condition known as nonalcoholic fatty liver disease (NAFLD) can range from moderate hepatic steatosis to nonalcoholic steatohepatitis (NASH) . NASH caGiven the analysis of individual MetS components, only high systolic blood pressure was significantly associated with high GSD prevalence and it raised the heterogeneity Figure 2NAFLD has traditionally been considered the hepatic manifestation of the metabolic syndrome because NAFLD is often associated with repertoire of MetS features . Leite eA growing number of studies suggested in multivariate adjustment analyses that gallstones were no longer independently associated with NAFLD but cholecystectomy was the independent risk factor for NAFLD , 42. ThoHigh heterogeneity can be caused by racial factors. According to our subgroup analysis and regression analysis, the phenomenon in Hispanics is about twice as common as than risks in Asians. Global figures showed that while NAFLD prevalence in Asia is only 27% , it is aIn fact, IR provides a key link between MetS, NAFLD, increased susceptibility to gallstones, and cholecystectomy . The corIt is important to note the reliability and usefulness of our findings. We evaluated the association between GSD and NAFLD/MetS by consideration of risk direction. To be specific, the pooled GSD risk in subjects with/without NAFLD/MetS, and risk of metabolic derangements based on GSD status were evaluated respectively in different models. Based on the positive linear dose-response relationship, using BMI to predict GSD has a high cost-effectiveness . In the We noted that there were certain limitations even though the majority of the included research were of high quality. First of all, we were unable to conclude with certainty that the observed outcomes were not attributable to NAFLD/MetS itself or to any of the other possible confounding variables. Second, only a small number of prospective cohort studies were included in each analysis, which limited the ability to obtain more conclusive evidence and the conclusions need to be confirmed in more and larger cohort studies. Third, it is necessary to state the drawbacks of combining data from different recruited research, such as the lack of consistency in risk measurements and statistical methodologies. RR and OR from different statistical models exhibited discrepant meaning, suggesting the presence of heterogeneities if the two were combined. Fourth, very few included studies could support the dose-response analysis of BMI, so the relationship between BMI and GSD would be compared later. Fifth, this study is not a mechanism study, so the potential mechanism of IR affecting GSD has not been well described. We will follow up on animal experiments to explore this process. Sixth, due to the lack of data on relevant risk factors provided in the registered articles, we cannot probe into the detailed biological interaction between NAFLD and BMI after cholecystectomy. Therefore, we are planning to collect more information based on information from our center to evaluate the quantitative relationship between cholecystectomy and NAFLD incidence in patients with different BMI categories , 101. In5This meta-analysis provided evidence that the close relationship between GSD and MetS/NAFLD, or insulin resistance, and the close relationship between cholecystectomy and NAFLD. No matter what kind of disease, geographical differences in the risks are greater in the America, compared to Europe and Asia. We also observed that calculating BMI might be a useful and customized technique for determining the likelihood of developing GSD. Paying attention to the control of blood pressure and blood sugar is helpful in alleviating GSD. In the future, well-designed and high-quality prospective studies are needed to confirm these effects and to further study cholecystolithotomy through metabonomics.The original contributions presented in the study are included in the article/ZL conceived and designed the study. JL and QL performed experiment and extracted information. JL and QL analyzed the data. JL wrote the manuscript. ZL reviewed the manuscript. ZL provided the funding support. All authors contributed to the article and approved the submitted version.This study is supported by Innovative Research Groups of National Natural Science Foundation of China (81721091), Major program of National Natural Science Foundation of China (91542205), National S&T Major Project (2017ZX10203205), National Natural Science Foundation of China (81902813), Zhejiang International Science and Technology Cooperation Project (2016C04003), Zhejiang Provincial Natural Science Foundation of China (LY22H030008), Zhejiang Medical Association (2019ZYC-A81), International Youth Exchange Programme by China Association for Science and Technology (2019), Tianqing Liver Diseases Research Fund (TQGB20200114), Organ Transplantation Overseas Training for Youth Talents from Shulan Excellent Talent Project, CSCO -Bayer Tumor Research Funding (Y-bayer202001/zb-0003), Chen Xiao-ping Foundation for the Development of Science and Technology of Hubei Province (CXPJJH122002-078), Beijing iGandan Foundation (1082022-RGG022), Open Fund of Key laboratory of High-Incidence-Tumor Prevention & Treatment , Ministry of Education, and Xinmiao Talent Supporting Program (2022R421A031).We profoundly grateful to our supervisor, ZL, whose illuminating instruction and expert advice have guided me through every step of my writing of this research. Our great gratitude also goes to some of our friends and classmates who have selfless and generously helped me with my thesis.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.


There have been a variety of published studies on the expression of serum miR-21 in patients with ovarian cancer associated with the diagnostic value of ovarian cancer, but the conclusions are not clearly elucidated. This study aims to evaluate the value of serum miR-21 expression in the diagnosis of patients with ovarian cancer by meta-analysis.Databases, such as PubMed, Embase, Web of Science, Cochrane Library, China National Knowledge Infrastructure (CNKI), and China WanFang, were searched for relevant studies upon the correlation between the expression of serum miR-21 and the diagnostic value of ovarian cancer from inception to March 7, 2022. Statistical analysis was performed using Stata 15.0 software. The pooled sensitivity, specificity, diagnostic odds ratio (DOR), positive likelihood ratio (PLR), and negative likelihood ratio (NLR) were calculated. The meta-regression analysis and subgroup analysis were used to explore the sources of heterogeneity. The Quality Assessment of Diagnostic Accuracy Studies-2 (QUADAS-2) system was used to evaluate the quality of the included literature.P > 0.05).A total of 6 articles were included in the meta-analysis. The results showed that the pooled sensitivity, specificity, PLR, NLR, and DOR were 0.81 (95%CI: 0.73–0.88), 0.82 (95%CI: 0.75–0.87), 4.51 (95%CI: 3.22–6.31), 0.23 (95%CI: 0.16–0.33), and 19.87 (95% CI: 11.27–35.03), respectively. The area under the summary receiver operating characteristic curve was 0.89 (95%CI: 0.85–0.91). No significant publication bias was found (Serum miR-21 has a good diagnostic value for ovarian cancer, which can be an ideal diagnostic biomarker for ovarian cancer. However, we should gingerly use miR-21 as a diagnostic reference standard, due to the limited number of included studies and heterogeneity. Ovarian cancer is a malignant tumor of the female reproductive system, which is characterized by its insidious onset and rapid progress, among which epithelial ovarian cancer (EOC) accounts for about 90%. According to the cancer statistics, 70–80% of ovarian cancer patients have been in the advanced stage before diagnosis, and the 5-year survival rate is only 20–30%, while the 5-year survival rate for initial-stage ovarian cancer is as high as 90% . Hence, In recent years, the role of miR-21 in tumor pathogenesis has been increasingly clarified, where miR-21 is involved in all known tumor biological processes, including tumorigenesis, development and metastasis –6. MultiIn order to obtain more accurate evidence to prove the relationship between the high expression of miR-21 and the diagnosis of ovarian cancer, this study conducted a comprehensive search of relevant literature and applied meta-analysis aiming to accurately assess the diagnostic value of serum miR-21 in ovarian cancer.We preformed this meta-analysis according to the guidelines of the preferred reporting items for systematic reviews and meta-analyses (PRISMA) .PubMed, Embase, Web of Science, Cochrane Library, China National Knowledge Infrastructure (CNKI), and China WanFang databases were searched about the studies on the diagnostic value of miR-21 for ovarian cancer. The retrieve time was from inception to March 7, 2022. The search strategy was as follows: (“microRNA-21” OR “miRNA-21” OR “microRNA21” OR “miR-21” OR “miRNA21”) AND . There was no language limitation. The titles and abstracts were checked by two researchers to identify all relevant trials, after then full text articles were scanned. The references of the relative review, and meta-analysis were also examined to identify the eligible studies. Two researchers searched each database independently, and finally cross-checked. If there were disagreements, they were resolved through discussion.Inclusion criteria: (1) studies aimed at miR-21 in serum samples of ovarian cancer; (2) studies focused on human samples; (3) studies were on the correlation between miR-21 and diagnostic accuracy; (4) studies provided data to extract or calculate true positive (TP), false positive (FP), false negative (FN), and true negative (TN).Exclusion criteria: (1) letters, case reports, reviews, conference abstracts, animal research; (2) diagnostic sensitivity and specificity could not be extracted; (3) repeated samples.The Quality Assessment of Diagnostic Accuracy Studies-2 (QUADAS-2) was applied for quality control of the included literature . The QUAThe following information upon the studies of the diagnostic value was extracted: (1) first author, year of publication, country, source of the sample, test method, number of patients, source of the control group, study duration, tumor stage, cut-off value, age; (2) sensitivity, specificity, the number of TP, FP, FN, TN. Data extraction was carried out independently by two researchers, disagreements resolved by discussion.I2) and P value were used to evaluate the heterogeneity among the studies [I2 > 50% or P> 0.05. The relationship between the prior probability, the likelihood ratio and the posterior test probability were evaluated by Fagan’s Nomogram plot [Stata 15.0 statistical software  was adopted to analyze the data and output the graph of bias risk assessment of the included studies. The diagnostic threshold effect was evaluated by the Spearman correlation between the logarithm of sensitivity and the logarithm of 1-specificity and the typical “shoulder-arm shape” in summary receiver operating characteristic (SROC) curve. We adopted the bivariate mixed-effects model to combine and analyze the data in our meta-analysis. The pooled sensitivity, specificity, positive likelihood ratio (PLR), negative likelihood ratio (NLR), and diagnostic odds ratio (DOR) were calculated, and the corresponding forest plot was drawn. The area under the curve (AUC) was obtained, where the AUC value ranged from 0.5 to 1.0. When it was close to 0.5, it indicated inadequate diagnostic performance. When it was close to 1.0, it indicated sound diagnostic performance. In addition, the inconsistency index ( studies . Heterogram plot . The metn = 524) and the control group (n = 486). The specific screening flow diagram is shown in Fig. A total of 708 articles were retrieved through the search strategy, of which the 373 repetitive articles were removed, 306 irrelevant articles were excluded through review of their titles and abstracts. After that, 23 articles were excluded. A total of 6 studies were included in this meta-analysis after checking full-text and data integrity –34. A toThe quality of the included literature was assessed following the QUADAS-2, shown in Fig. P > 0.05), indicating no threshold effect among the diagnostic test. The DOR  , specificity  , and NLR  , which all may be the source of heterogeneity  was an unfavorable prognostic factor in the multivariate cox-regression model [Histopathological examination is the gold standard for the diagnosis of malignant tumors, whereas it has not been widely applied in cancer diagnosis because of its invasiveness and high cost . At preson model . Besideson model . Multiplon model –47. The on model . HoweverThe pooled sensitivity and specificity of the six studies were 0.81 and 0.82, respectively, which suggested that miR-21 has sound sensitivity and specificity in the diagnosis of ovarian cancer. The PLR and NLR were 4.51 and 0.23, respectively, indicating that the possibility of correct judgment of positive in diagnostic test was 4.51 times higher than that of misjudgment, while the possibility of misjudging negative was 23% of that of correct judgment. It is reported that the PLR should be more than 10 and the NLR should be less than 0.1 to ensure convincing accuracy , which sThis study has found that the heterogeneity of the pooled effect is not caused by the threshold effect, which means heterogeneity results from other factors. It was meta-regression analysis that discovered such factors as different sample sizes of ovarian cancer patients, different national populations and different cut-off values were probably important sources of heterogeneity. The reason might be that the demographic characteristics varied among different regions, and so did the detection rate of positive results under different sample sizes and different cut-off values. The subgroup analysis showed that the diagnostic value of Egypt group, control group, sample size ≤80 group and cut-off value ≤1.5 group were higher than that of the corresponding groups. The result is significant in guiding further data collection of the follow-up research, and more than that, it provides a subgroup basis for the subsequent subgroup analysis. The results of sensitivity analysis show that the results of this study are stable. No significant publication bias was found in this meta-analysis.Indeed, this meta-analysis was limited by the following factors. Firstly, the time when miR-21 was used as a molecular marker of tumor was short, with relatively a few samples included. Accordingly, more studies are needed with larger sample size. Secondly, early detection of ovarian cancer means a lot to clinically significance. In this study only one article discussed the diagnostic value of miR-21 in early ovarian cancer , two artIn conclusion, serum miR-21 has a good diagnostic value in ovarian cancer, which we can consider combining with other biomarkers or other diagnostic methods to improve the diagnostic accuracy of patients with ovarian cancer. Given the heterogeneity of the research, more samples with larger sizes are still needed to verify the findings prospectively in the future.


Results: Individuals with ASD showed a significantly higher WBC count , significantly higher levels of neutrophils , monocytes , NK cells , Tc cells , and a significantly lower Th/Tc cells ratio , compared to TD controls. Subjects with ASD were also characterized by a significantly higher neutrophil-to-lymphocyte ratio (NLR) , and significantly higher neopterin levels  compared to TD controls. No significant differences were found with respect to the levels of lymphocytes, B cells, Th cells, Treg cells, and Th17 cells. Sensitivity analysis suggested that the findings for monocyte and neutrophil levels were robust, and independent of other factors, such as medication status, diagnostic criteria applied, and/or the difference in age or sex between subjects with ASD and TD controls. Taken together, our findings suggest the existence of a chronically  activated inflammatory response system in, at least, a subgroup of individuals with ASD. This might have not only diagnostic, but also, therapeutic implications. However, larger longitudinal studies including more homogeneous samples and laboratory assessment methods and recording potential confounding factors such as body mass index, or the presence of comorbid psychiatric and/or medical conditions are urgently needed to confirm the findings. Autism spectrum disorder (ASD) is a neurodevelopmental condition with a so far unknown etiology. Increasing evidence suggests that a state of systemic low-grade inflammation may be involved in the pathophysiology of this condition. However, studies investigating peripheral blood levels of immune cells, and/or of immune cell activation markers such as neopterin are lacking and have provided mixed findings. We performed a systematic review and meta-analysis of studies comparing total and differential white blood cell (WBC) counts, blood levels of lymphocyte subpopulations and of neopterin between individuals with ASD and typically developing (TD) controls (PROSPERO registration number: CRD CRD42019146472). Online searches covered publications from 1 January 1994 until 1 March 2022. Out of 1170 publication records identified, 25 studies were finally included. Random-effects meta-analyses were carried out, and sensitivity analyses were performed to control for potential moderators.  HLA) might be associated with ASD [Autism spectrum disorder (ASD) is considered a neurodevelopmental condition characterized by persistent deficits in social communication and by restrictive, repetitive and/or stereotyped behaviors . With a with ASD ,6, and iwith ASD . Moreovewith ASD . Howeverwith ASD ,10, and with ASD  in theirwith ASD , turningwith ASD , would bThis systematic review and meta-analysis were undertaken and reported according to the Preferred Reporting Items for Systematic Reviews and Meta-analyses (PRISMA) guidelines . The PRIIn-and exclusion criteria were discussed and approved by all authors. The following inclusion criteria were applied: (1) English-, Spanish, German, and/or French-written observational studies showing comparative data on total and differential white blood cell (WBC) counts, or on blood levels of lymphocyte subpopulations or neopterin between individuals with ASD and TD healthy controls, (2) ASD diagnosis according to DSM-IV, DSM-IV-TR, DSM-5 and/or ICD-10 criteria, (3) necessary data available , standard error (SE), range or interquartile range (IQR) for the levels of immune/inflammatory markers, as well as the numbers of individuals with ASD and TD healthy controls). Excluded were studies meeting at least one of the following criteria: (1) non-original studies , (2) non peer-reviewed studies , patents, safety and/or non-comparative studies , (3) animal model studies, (4) studies focusing on certain tissue/cerebrospinal fluid instead of peripheral blood, (5) studies assessing genetic polymorphisms, (6) targeted sampling strategies , (7) individuals with Rett’s syndrome, (8) significant study overlap , and (9) studies using siblings as controls. Grey literature was not included. PubMed, SCOPUS, and World of Knowledge electronic databases were independently searched by two authors  for relevant articles published between 1 January 1994 and 1 March 2022. The following search syntax  was used for PubMed database search, and adapted according to the different database index terms: (autism OR ASD OR autistic* OR asperger*) AND . Title and abstracts of all electronic articles retrieved were independently screened by two authors . Reference lists of included studies and relevant topic-related reviews were manually screened to retrieve any additional reports meeting criteria. Articles presumed to meet the inclusion criteria were retrieved as full-texts, and independently examined by the same authors, reaching a preliminary list of included studies. In case of disagreement, a consensus including all authors was reached. A final list of included articles was then reached. A flow-chart summarizing the study selection process is presented in One author (GA-H) extracted all data, the method used to extract the data was independently verified by two other authors  and data were re-checked for accuracy by a third author (JL-M). Data extracted included: name of first author, date of publication, country of origin, population description , medication (yes/no/not given), diagnostic criteria to establish a diagnosis of ASD, total and differential WBC and blood levels of lymphocyte subpopulations and/or neopterin. Whenever possible, data were retrieved as mean and standard deviation (SD). Conversion methods were used to calculate mean and SD in case of data provided as median, range, interquartile range (IQR), and standard error (SE). The median was included as an approximation of the mean ; the SD was calculated using the following formulas: SD = IQR/1.35; SD = SE × √N, or SD = maximum-minimum/4 ,16.CorreTwo researchers  independently evaluated the quality of included studies according to the Newcastle-Ottawa Scale (NOS) . The NOSI2 statistic; heterogeneity was considered moderate when I2 ranged between 50% and 75%, and high when I2 was greater than 75% [p-values were considered statistically significant at the α = 0.05 level. Differences in blood levels of immune/inflammatory markers between individuals with ASD and TD controls were tested as main outcome measures. Meta-analyses were conducted when two or more studies were included using a similar modality of biomarker assessment. When multiple studies investigated the same sample, analysis included only the study with the largest sample size. Effect size (ES) estimates were based on Hedge’s g and their 95% CI for every individual marker, considering Hedge’s values from 0.2 to 0.49 small, values from 0.5 to 0.79 moderate, and greater than 0.8, large. Based on the expected between-study differences in sample size and/or recruitment procedures and methods used to determine the levels of assessed markers, random-effects models were used . A positthan 75% . Publicathan 75% . The Orwthan 75% . Data anOnline searches identified 1770 publication records; 25 studies ,44,45,46g = 0.29, 95% CI: 0.13–0.45, p = 0.001) (I2 = 34%)  or to DSM-5 criteria . On the contrary, sensitivity subgroup meta-analyses appeared not significant when the analysis was limited to studies based on medication-naïve individuals with ASD . Meta-regression analyses showed that the ES for the difference in the WBC count between individuals with ASD and TD were not significantly associated with moderators age, sex, and/or year of publication  . Heterog2 = 34%) . After r2 = 34%) . Results2 = 34%) . Sensitilication .g = 0.29, 95% CI: 0.08–0.49, p = 0.005) (I2 = 31%)  . Heterog2 = 31%) . After r2 = 31%) . Meta-re2 = 31%) . g = 0.35, 95% CI: 0.17–0.54, p < 0.001) (I2 = 54%)  , or to DSM-5 criteria . Meta-regression analyses showed that the ES for the difference in monocyte levels between individuals with ASD and TD was not significantly associated with age, sex, and/or year of publication  . Heterog2 = 54%) . After r2 = 54%) . Results2 = 54%) . Sensiti2 = 64%) . The diflication .g = −0.055, 95% CI: −0.229–0.119, p = 0.533, I2 = 24%) , or to studies based on participants with an ASD diagnosed according to DSM-IV/ICD-10 criteria , or to DSM-5 criteria . Meta-regression analyses showed that the ES for the difference in the monocyte levels between individuals with ASD and TD was not significantly associated with moderators age, sex, and/or year of publication  . After r2 = 24%) . The diflication . g = −0.09, 95% CI: −0.58–0.40, p = 0.722, I2 = 80%) . Sensitivity subgroup meta-analyses based on studies where individuals with ASD were diagnosed according to DSM-IV/ICD-10 criteria confirmed the direction, magnitude and significance of the associations for B cell levels . Due to the low number of studies, subgroup analyses based on participants with ASD diagnosed according to DSM-5 criteria were not performed, and potential moderators were not tested. The levels of B cells were assessed by five studies ,28,29,342 = 80%) . After r2 = 80%) . Sensitig = 0.36, 95% CI: 0.02–0.71, p = 0.037) (I2 = 67%) , on participants with ASD diagnosed according to DSM-IV/ICD-10 criteria , and/or participants with ASD diagnosed according to DSM-5 criteria . Meta-regression analyses demonstrated that between-group differences in age, sex, and year of publication were not significantly associated with ES estimates for NK cell levels  . Heterog2 = 67%) . After r2 = 67%) . Subgroul levels .g = 0.024, 95% CI: −0.367–0.414, p = 0.905, I2 = 68%) , and on participants with ASD diagnosed according to DSM-IV/ICD-10 . Due to the low number of studies, subgroup analysis based on participants with ASD diagnosed according to DSM-5 criteria were not performed, and potential moderators were not tested.T cell levels were assessed by five studies ,28,29,342 = 68%) . After r2 = 68%) . Subgroug = 0.726, 95% CI: 0.111–1.342, p = 0.021) , and on participants with ASD diagnosed according to DSM-IV/ICD-10 . Due to the low number of studies, subgroup analysis based on participants with an ASD diagnosed according to DSM-5 criteria could not be performed; potential moderators were also not tested.The levels of Tc cells were assessed by four studies ,27,29,34= 0.021) . Heterog= 0.021) . Subgroug = −0.35, 95% CI: −1.10–0.41, p = 0.370) , and on participants with ASD diagnosed according to DSM-IV/ICD-10 . Due to the low number of studies, subgroup analysis based on participants with ASD diagnosed according to DSM-5 criteria could not be performed, and potential moderators were not tested.The levels of Th cells were assessed by five studies ,29,34,36= 0.370) . Heterog= 0.370) . Subgroug = −0.42, 95% CI: −0.73–−0.11, p = 0.008) (I2 = 0%). Sensitivity analyses were not performed. The Th/Tc cells ratio was assessed by three studies ,29,34. O= 0.008) ; heterogg = 1.75, 95% CI: −0.25–4.37, p = 0.081) (I2 = 97%) . Individuals with ASD were diagnosed according to DSM-IV/ICD-10 criteria in only one study [The levels of Th17 cells were assessed by three studies ,33,40. O= 0.081) . Heterog2 = 97%) ; after r2 = 97%) . The difne study ; medicatg = −1.30, 95% CI: −2.61–0.04, p = 0.051) (I2 = 97%) . The difference remained not significant when the analysis was limited to studies based on medication-naïve participants , and on participants with an ASD diagnosis based on DSM-IV/ICD-10 criteria . Meta-regression analyses demonstrated that between-group differences in age, sex, and/or year of publication were not significantly associated with ES estimates for Treg cell levels  . Heterog2 = 97%) . After r2 = 97%) . Subgroul levels . g = 0.69, 95% CI: 0.03–1.34, p = 0.040) (I2 = 90%) . Due to the low number of studies, potential moderators were not tested.The NLR was assessed by four studies ,39,42,46= 0.040) . Heterog2 = 90%) . After r2 = 90%) . Subgroug = 1.16, 95% CI: 0.62–1.69, p = 0.001) (I2 = 72%)  . Heterog2 = 72%) . After r2 = 72%) . Due to +) cells [To the best of our knowledge, this is the largest meta-analysis to date exploring differences in blood peripheral levels of immune cells and immune cell activation markers  between individuals with ASD and TD controls. Overall, 25 studies were included. We found that subjects with ASD were characterized by a significantly higher WBC count, and by significantly higher neutrophil and monocyte blood levels, compared to TD controls. ES was low for all the above-mentioned immune markers. Heterogeneity was low for the WBC count and for neutrophil levels, and moderate for monocyte levels. Sensitivity analyses suggested that the findings related to neutrophil and monocyte levels were robust, and independent of other factors, such as medication status, diagnostic criteria applied, year of publication, and/or the differences between subjects with ASD and TD in relation to age and/or sex. The WBC consists of different cell types  and it is currently considered as a reliable measure of the overall immune system activity. In other words, a high WBC count  suggests the existence of an activated response system (IRS) . While t+) cells . In addi+) cells , increas+) cells , support+) cells , suggest+) cells ,53. In a+) cells . Since a+) cells ,56, we hOur findings might have diagnostic and therapeutic implications. A significant positive association between the NLR and/or blood levels of neopterin, and the severity of ASD  have been reported in individuals with ASD ,39,46. OAlthough encouraging, our findings should be interpreted with caution considering several limitations. First, this study has focused on peripheral immune cell and neopterin levels, making it possible that outcomes differ from measurements in the brain or in the CSF. Second, high levels of between-study heterogeneity were recorded for most several immune/inflammatory markers  assessed in our analysis. While the meta-analytic approach attempts to adjust for methodological confounders, the procedure was limited by the number of available studies and the information that could be extracted from the included studies. Unexplained heterogeneity could be related to potential moderators that were not consistently addressed in the different studies, such as BMI  and/or tTaken together, findings from this meta-analysis suggest that at least a subgroup of subjects with ASD might show abnormal levels of immune cells  and increased peripheral levels of the immune cell activation marker neopterin, compared to TD controls. These findings suggest the existence of a (chronic) activated IRS in ASD. In particular, individuals with ASD might be characterized by a chronic activation of the monocyte/macrophage system, and by abnormal blood levels of different lymphocyte subpopulations. Stratification of individuals with ASD based on (immune/inflammatory) parameters could be a method for improving diagnosis and increasing the effectiveness of interventions . However, rigorously designed and larger longitudinal studies, including more homogeneous samples and recording potential moderators such medication status, BMI, and/or the presence of comorbid mental and/or medical disorders are urgently needed to confirm our findings.


Biliary tract cancers (BTCs), encompassing cholangiocarcinoma (CCA), gallbladder (GBC), and ampulla of Vater cancers (AVC), are common hepatobiliary cancer after hepatocellular carcinoma with a high mortality rate. As there is no effective chemopreventive agent to prevent BTCs, this study aimed to explore the role of statins on the risk of BTCs.PubMed, Embase, and Cochrane Library from inception until 24 April 2020 were searched according to the Meta‐Analyses of Observational Studies in Epidemiology (MOOSE) guidelines. The adjusted risk ratios (aRRs) of BTCs and individual cancer were pooled using a random‐effects model.Eight observational studies (3 cohort and 5 case–control studies) were included with 10,485,231 patients. The median age was 68.0 years (IQR: 67.0–71.5) and 48.3% were male. Statins were associated with a lower risk of all BTCs . The pooled aRR for CCA was 0.60 (95% CI: 0.38–0.94) and GBC was 0.78 (95% CI: 0.68–0.90). There was only one study on AVC with aRR of 0.96 (95% CI: 0.66–1.41). The pooled aRR for lipophilic and hydrophilic statins was 0.78 (95% CI: 0.69–0.88) and 0.70 (95% CI: 0.61–0.80), respectively. The effects were attenuated in studies that adjusted for aspirin and/or non‐steroidal anti‐inflammatory drugs  and metformin .Statins, both lipophilic and hydrophobic, were associated with a lower risk of BTCs, particularly CCA and GBC. Statins associate with lower risk of biliary tract cancers: a systematic review and meta‐analysis. The inclusion criteria included2.3For the eligible studies, items including first authors, publication year, country, study design, inclusion/exclusion criteria, exposure of interest, outcome of interest, sample size, age, sex, follow‐up time, and variables considered in the propensity score (PS) analysis or multivariable analysis were recorded. The PS represents the probability of prescribing statins that is dependent on other covariates . PS matching ensures the balance of measured confounding factors between statins and control groups, so that any difference in the outcome (BTCs) would ideally be due to the effect of statin use only. We also recorded data of adjusted effect estimates (aHR/aOR/adjusted relative risk) derived by PS matching or multivariable analysis. The quality of observational studies was assessed by two authors (MYWY and WSW) independently using the Newcastle‐Ottawa scale (NOS) with the highest score of nine points. The risk of bias was categorized into 3 groups: Low risk (7–9 points), moderate risk (4–6 points), and high risk (<4 points).2.4All statistical analyses were performed using the R version 3.2.3  statistical software. Continuous variables were expressed as median (interquartile range [IQR]) or mean (±1 standard deviation [SD]). Comparisons of outcome were expressed as adjusted risk ratio (aRR) with 95% confidence interval (95% CI) using the random‐effects model, and were presented as Forest plot. Studies that did not report adjusted effect estimates were excluded from the primary analysis since striking imbalance in baseline characteristics likely existed between statin and statin non‐users. As statins users more often have cardiovascular risk factors including DM, smoking, alcohol use which are also known risk factors for BTCs,p‐value < 0.1 indicating significant heterogeneity. We calculate I2 statistics to measure the proportion of total variation in study estimates attributed to heterogeneity. I2 values of <25%, 25%–75%, and >75% indicate low, moderate, and high heterogeneity, respectively.We used Cochran Q test to detect heterogeneity among studies, with a The effect estimate of dose of statins on the BTC was also calculated. There were two studiesThe pooled effect estimate from unadjusted RR was also reported. Subgroup analysis was performed according to the class of statins (lipophilic vs. hydrophilic), cancer type, ethnicity (East vs. West), study design (case–control study vs cohort study), presence of adjustment for aspirin and/or non‐steroidal anti‐inflammatory drugs (NSAIDs), metformin, smoking, alcohol use, DM, and PSC. Lipophilic statins that were investigated in various studies included atorvastatin, simvastatin, lovastatin, fluvastatin, while hydrophilic statins included rosuvastatin and pravastatin. Statin users often had underlying cardiovascular diseases and risk factors necessitating concomitant aspirin and/or metformin use, and these two drugs have been reported to be associated with lower BTC risk.3Figure 3.1A total of 10,485,231 subjects were included in the primary analyses. The median age of patients was 68.0 years (IQR: 67.0–71.5) and 48.3% were male. The characteristics of comparison group are shown in Table p < 0.001; I2 = 92.46%). Funnel plot did not suggest publication bias (Egger's test: p = 0.626)   Figure . There w) Figure . Five st) Figure .3.2p = 0.007, I2 = 61.15%). Three studies were on atorvastatin with an aRR of 0.77 (95% CI: 0.61–0.98), three on simvastatin with an aRR of 0.85 (95% CI: 0.65–1.13), one on lovastatin with an aRR of 0.69 (95% CI: 0.58–0.83), and one on fluvastatin with an aRR of 0.77 (95% CI: 0.61–0.96) . There were two studies on rosuvastatin with an aRR of 0.79 (95% CI: 0.53–1.18), and two on pravastatin with an aRR of 0.77 (95% CI: 0.55–1.07)  Table . For hyd6) Table . There w7) Table .3.3There were two studies that reported a dose–response relationship between statins and BTCs.3.43.4.1p < 0.001, I2 = 93.52%) . There were two studies on eCCA, with aRR being 0.37 (95% CI: 0.13–1.02) and there was no significant heterogeneity .Five studies on CCA were included. The pooled aRR was 0.60 (95% CI: 0.38–0.94), with significant heterogeneity among the studies  Figure . There w3.4.2p = 0.353, I2 = 16.71%) . There was no significant heterogeneity between the studies  Figure .3.4.3There was only one study investigating AVC, reporting an aRR of 0.96 (95% CI: 0.66–1.41).3.4.4p = 0.282, I2 = 11.22%)   and there was no significant heterogeneity  Figure . For the) Figure .3.4.5p < 0.001, I2 = 90.96%)  with pooled aRR of 0.65 (95% CI: 0.48–0.88). There was significant heterogeneity among the studies  Figure . There w3.4.6Three studies adjusted for aspirin and/or non‐aspirin NSAID use, with a pooled aRR of 0.80 (95% CI: 0.72–0.89) Figure A, while Two studies adjusted for metformin use, with a pooled aRR of 0.80 (95% CI = 0.72–0.90) Figure . The remThere were five studies that adjusted for smoking with a pooled aRR of 0.62 (95% CI: 0.34–1.11) Figure . For theFour studies adjusted for alcohol use showed a pooled aRR of 0.63 (95% CI: 0.30–1.30) Figure , while tSix studies adjusted for DM with a pooled aRR of 0.64 (95% CI: 0.40–1.03) Figure . For theTwo studies adjusted for PSC, with a pooled aRR of 0.75 (95% CI: 0.57–0.99) Figure . The rem4In this meta‐analysis of eight studies involving more than 10 million patients, we found that statins use was associated with a 33% lower risk of BTCs, including CCA and GBC. Both lipophilic and hydrophobic statins were associated with lower risks of BTCs.While the main indications of statins are hyperlipidemia and prevention of major vascular events such as stroke or acute coronary syndrome, statins were also found to possess anti‐inflammatory and immunomodulatory effectsThe chemopreventive effects of statins can be explained by the inhibitory effects of statins on cell proliferation via multiple mechanisms. First, statins suppress cell migration and invasion by inhibiting post‐translational modification of G‐proteins that are involved in growth factor signal cascades, cell membrane trafficking, and cell adhesion, such as Ras, Rac, and Rho.Pooling adjusted effect estimates is important in assessing the potential association between statins and cancer development. One may argue that the observed beneficial effect of statins is due to healthy user bias. However, the median age of included subjects in this meta‐analysis was 68.0 years and older individuals (in whom BTCs more likely to occur) generally have more comorbidities that may increase cancer risk, hence potentially negating the concern of healthy user bias.In‐vitro studies suggest a differential chemopreventive effect between hydrophilic and hydrophobic statins—lipophilic statins are associated with cytotoxicity in cancer cellsn = 3). Moreover, observational studies with adjustment for different variables makes it unreliable to compare the effect of individual statins with each other.Furthermore, the mechanisms of each type of statin vary. For instance, simvastatin acts on dysregulating the cholesterol biosynthesis pathway, while fluvastatin depends on reactive oxygen species formation. Nevertheless, we found that both lipophilic and hydrophilic statins were associated with a lower risk of BTCs . Although network meta‐analysis can characterize the comparative efficacy of various statins by indirect comparison, it may not be ideal in this scenario because the scanty number of studies on individual statin effect , ethnicity (West vs. East), and presence of adjustment for certain variables . The pooled effect estimate was similar for cohort and case–control study design (0.67 and 0.60 respectively), although it was of borderline significance for case–control study design. While the pooled effect estimate was 0.65 and 0.80 for studies conducted in West and East respectively, more studies are warranted from the Asian population as there was only one study from Taiwan. The study from Taiwan was the only one that adjusted for parasitic infection. Importantly, the protective effective was lower in studies which adjusted for aspirin (20% vs. 38% risk reduction) and metformin (20% vs. 37% risk reduction). Aspirin reduces CCA risk by suppressing bile duct inflammation and cancer growth factor pathways through COX‐2 inhibition,A few limitations of this study should be acknowledged. First, there are so far no randomized controlled trials (RCTs) investigating the chemopreventive role of statins in BTCs development. All included studies are observational studies, hence raising the possibility of unmeasured/residual confounding. In particular, case–controls studies generally suffer more biases than cohort studies, including selection bias, recall bias, and interviewer bias. Second, a significant majority of the study subjects were aged ≥60 years, and therefore whether statins are protective among younger subjects requires further investigation. Third, a number of risk factors for BTCs, such as cholelithiasis, PSC, bile duct cysts or parasitic infections,5Our meta‐analysis showed that statins were associated with a 33% lower risk of BTCs, in particular CCA and GBC. The beneficial effect was observed for both lipophilic and hydrophilic statins, and was consistent with cohort study or case–control study. Future prospective RCTs are warranted to confirm the chemopreventive effect of statins on BTCs development.Ka Shing Cheung was involved with study concept and design, analysis and interpretation of data, drafting of manuscript, and approval of the final version of the manuscript. Matthew YW Yeung and Wing Sum Wong was involved with literature search, analysis and interpretation of data, and drafting of manuscript. Bofei Li was involved with analysis and interpretation of data and drafting of manuscript. WK Seto was involved in interpretation of data and critical revision of the manuscript. Wai K Leung was involved with the study concept and design; analysis and interpretation of data; critical revision of the manuscript for important intellectual content; study supervision; and approval of the final version of the manuscript.KSC has received speaker fee from AstraZeneca and honorarium for attending advisory board for Janssen and AstraZeneca. WK Seto has received speaker fee from AstraZeneca and Mylan, honorarium for attending advisory board for CSL Behring, and speaker fee and honorarium for attending advisory board for AbbVie and Gilead. WKL has received speaker fee from Eisai, Ferring, Ipsen, and honorarium for attending advisory board for Janssen and Pfizer.Ethics approval was not required as this was a meta‐analysis of published articles.Patient consent was not required as this was a meta‐analysis of published articles.Appendix S1Click here for additional data file.


FXIII polymorphisms. However, studies on the role of the Val34Leu polymorphism in the aetiology of vascular diseases often show contradictory results. In the present meta-analysis, we aimed to pool data from available articles to assess the relationship between the FXIII Val34Leu polymorphism and the susceptibilities to IS of undetermined source and premature MI in patients aged below 55 years. Methods: We searched databases  using specific keywords (the last search was in January 2022). Eventually, 18 studies  met the inclusion criteria. Data were analysed using RevMan 5.4 and StatsDirect 3 link software. The relation between Val34Leu polymorphism and disease was analysed in five genetic models, i.e., dominant, recessive, additive, heterozygous, and allelic. Results: No relation between Val34Leu polymorphism and IS in young adults was observed in all analysed genetic models. For premature MI, significant pooled OR was found between the carrier state of the Leu allele  and a lack of MI, suggesting its protective role . A similar finding was observed for the heterozygous model in MI  . No relation was found for the recessive, additive, and allelic models in MI. Conclusions: In the population of young adults, no positive correlation was found between the FXIII Val34Leu polymorphism and IS of undetermined source in any of the analysed genetic models. In turn, the carrier state of the 34Leu allele as well as FXIII heterozygotes themselves were found to play a protective role in relation to premature MI.Background: Polymorphisms within the gene that encodes for coagulation factor XIII (FXIII) have been suggested to be involved in the pathogeneses of ischemic stroke (IS) and myocardial infarction (MI). The Val34Leu polymorphism is one of the most commonly analysed  The contribution of both genetic predisposition and well-known conventional risk factors to the pathogenesis of the first cardiovascular (CV) event in young people appears to be obvious.Factor XIII (FXIII) is a multifunctional pro-γ-transglutaminase involved in the formation and stabilisation of a fibrin clot as well as the aggregation and adhesion of platelets ,2. PreviFXIII gene polymorphism, G > T transition in the second exon of the F13A1 gene, results in the substitutive exchange of leucine (Leu) for valine  in the A subunit. Research has found higher activity of FXIII in Leu carriers, while Val homozygotes present a decrease in the activity of this factor [FXIII also influences the structure of a fibrin clot, especially in the presence of increased concentrations of fibrinogen, making it much tighter [The common s factor ,7. The V tighter . The rol tighter ,11,12,13Cryptogenic stroke (CS), possibly embolic, is a common type of IS in the young population, accounting for up to 60% of stroke cases before age 45 and approximately 25% in the 45–49 age range . Its nonIn turn, the most common cause of premature MI or IS in people with unfavourable family history and/or conventional atherogenic risk factors is accelerated atherogenesis and atherothrombosis . AtherosFXIII Val34Leu polymorphism and premature CV events of atherosclerotic or thrombotic origin in the population of young adults.The aim of the present study was to summarise the results of available data regarding the relationship between FXIII Val34Leu polymorphism related to IS and MI susceptibility when comparing the prevalence of its alleles and genotypes in young patients with IS and MI against that in controls, according to five genetic models?We searched five databases  to identify available data published before January 2022 with the use of appropriate keywords:  and  and (“young adults” or “young” or “premature” or “early”). The identified studies were included in accordance with the population, intervention, comparison, and outcome (PICO) model to select the relevant research question: Is Searched studies were included in the meta-analysis if: (a) there was confirmed ischemic stroke or myocardial infarction, (b) a case–control study methodology was used, (c) the age of the patients was below 55 years, (d) access to genotypes distribution was available, (e) the article was a full-length paper or brief communication, and (f) the article was written in English. Studies were excluded from the meta-analysis for the following reasons: (a) unavailability of genotyping results, (b) lack of reference (control) group, (c) lack of information on the age of the patients, or if the age of the patients was above 55 years, (d) if the material took the form of conference proceedings, review articles, case reports, or meta-analyses, or if the study was an animal study, and (e) if the article was written in a language other than English. When subgroups of patients younger than age 55 were available in the included studies, we used the data regarding those patients.FXIII polymorphism with regards to IS and MI met the inclusion criteria , including 6 studies on IS  to check the consistency of genotype distribution at the significance level of 0.05 for controls in each study that was included. The Newcastle–Ottawa scale (NOS) for case–control studies was used to evaluate the methodological quality of the included studies . Using t2 test, which describes the proportion of variance (from 0% to 100%) due to variance in true effect sizes rather than sampling error. I2 values of 25%, 50%, and 75% were correlated with low, intermediate, and high inconsistency, respectively. The random effects method  was used to calculate the pooled OR with a 95% CI when heterogeneity between the studies was significant; otherwise, the calculation was performed with the fixed-effects method . The strength of the correlation between the FXIII Val34Leu polymorphism and IS and MI was assessed in the following models: dominant , recessive , additive , heterozygous , and allelic . To evaluate the stability of the results, sensitivity analyses were made via the sequential exclusion of each study.Statistical analyses were conducted twice using the Review Manager software  and StatsDirect 3 link software . We calculated the pooled odds ratio (OR) with a 95% confidence interval (CI) to determine the strength of association between the particular genetic model and the disease, that is, IS or MI. We selected the statistical model of the analyses (random or fixed) based on heterogeneity between the included studies; this was assessed using the Ip value was below 0.05.To assess potential publication bias, both Egger’s regression and Begg’s rank correlation tests were performed. The result was considered statistically significant if the FXIII gene and ischemic stroke in young patients are shown in Characteristics of the six included studies analysing Val34Leu polymorphism within the FXIII polymorphism , no relation between the carrier state of the Leu allele and IS in young adults was observed. Similar findings were observed for the recessive , additive , and heterozygous model , as well as for the allelic  models  models .In the sensitivity analysis, no change in the OR value was demonstrated in the case of all analysed genetic models after excluding subsequent studies. Therefore, these analyses were considered stable.Regarding the analyses for IS, publication bias was observed for the dominant and heterozygous genetic models. For the remaining models, no publication bias was observed since the shapes of the funnel plots were roughly symmetrical. FXIII gene and premature MI are shown in FXIII Val34Leu polymorphism (in 7 out of 13 studies) [Characteristics of the thirteen studies analysing Val34Leu polymorphism within the studies) ,31,32,33studies) ; and Siestudies) , whereasstudies) , Butt etstudies) .FXIII polymorphism , significant pooled OR was demonstrated between the carrier state of the Leu allele and a lack of MI, suggesting its protective role . A similar finding was observed for the heterozygous model  could have a protective role. The carrier state of the Leu allele was more common in controls compared to young patients with MI . Subjects with Val/Leu genotypes were more frequent in controls than in MI patients  in reference to wild-type homozygous Val/Val, which may also suggest a protective effect. However, these results should be treated with caution since there was some loss of significance after omitting subsequent studies. The results for the remaining genetic models did not reveal significance between FXIII polymorphism and premature MI.The results of the present meta-analysis show different relationships for two types of ischemic events with different pathogeneses. In the case of IS with a cryptogenic background, often secondary to thromboembolic processes, we observed no relation with FXIII polymorphisms, have been analysed in the context of premature CV events, including coronary artery disease (CAD) [Coagulation factor polymorphisms, including se (CAD) ,41, IS [se (CAD) , haemorrse (CAD) , and vense (CAD)  in varioNumerous data confirmed the protective effect of 34Leu allele carriage on the development of premature MI. This effect was stronger in the 18–50-year-old population than in patients over 50 years of age ,35,36,45Val34Leu polymorphism is characterised by high ethnic variability. The prevalence of the 34Leu allele in Caucasians has been estimated at 37–51% ,36,42,44FXIII 34Leu variant (OR = 1.1). In a group of 1030 Turkish patients, the protective role of the FXIII Val34Leu polymorphism against MI was confirmed (OR = 0.31), but it was not an independent variable when major CV risk factors were taken into account in multivariate analysis [In 1210 young adult Italian people  with a hanalysis .FXIII 34Leu variant significantly reduced the risk of MI among smokers . In the above study, the risk reduction was greater in homozygotes than in heterozygotes for the Leu allele, suggesting a gene dosage effect.Franco et al.  reportedImportantly, the meta-analysis by Jung et al.  found thConversely, most researchers did not confirm the protective effect of the Val34Leu polymorphism in relation to IS ,55,56,57Elbaz et al.  describeAn additional problem that makes it difficult to confidently assess the contribution of gene polymorphisms in CS patients is the presence of undetected, asymptomatic atrial tachyarrhythmias such as atrial fibrillation and atrial flutter. Large clinical trials ,62 have FXIII Leu34 allele and prothrombin 20210G > A (FII 20210A) coexisted. In turn, a decrease in the risk of MI was found in 34Leu carriers with high fibrinogen levels [It has also been proven that the interactions between various genetic factors and gene polymorphisms involved in the coagulation and fibrinolysis processes are important in the pathogeneses of premature atherosclerosis and hypercoagulability, and that their synergistic effect may be crucial in the development of CAD/MI ,63 and In levels ,51. Reinn levels  reportedn levels , the effn levels  and IS [n levels ,67.FXIII polymorphism in the development of CV were available. Meta-analyses of some specific interactions between particular genes and factors that are simultaneously present in patients would be more accurate for understanding the role of the analysed polymorphism and the disease. In the case of FXIII Val34Leu polymorphism, the level of factor XIII should be especially considered.The present meta-analysis has some limitations. No additional data on other factors that could interact with the analysed FXIII Val34Leu polymorphism and IS in any of the analysed genetic models. In the case of premature MI, our meta-analysis demonstrated that the carrier state of the 34Leu allele might play a protective role in premature MI. In both cases, the influence of gene–gene and gene–environment interactions on disease development should be taken into account.The assessment of gene polymorphisms involved in the processes of coagulation and fibrinolysis may be important in the primary prevention of cardiovascular events in a group of young adults without classic risk factors. In young adults, no positive correlation was found between the


Changes in phenology and distribution are being widely reported for many migratory species in response to shifting environmental conditions. Understanding these changes and the situations in which they occur can be aided by understanding consistent individual differences in phenology and distribution and the situations in which consistency varies in strength or detectability.Studies tracking the same individuals over consecutive years are increasingly reporting migratory timings to be a repeatable trait, suggesting that flexible individual responses to environmental conditions may contribute little to population‐level changes in phenology and distribution. However, how this varies across species and sexes, across the annual cycle and in relation to study  and/or ecosystem characteristics is not yet clear.Here, we take advantage of the growing number of publications in movement ecology to perform a phylogenetic multilevel meta‐analysis of repeatability estimates for avian migratory timings to investigate these questions. Of 2,433 reviewed studies, 54 contained suitable information for meta‐analysis, resulting in 177 effect sizes from 47 species.Individual repeatability of avian migratory timings averaged 0.414  across landbirds, waterbirds and seabirds, suggesting consistent individual differences in migratory timings is a common feature of migratory systems. Timing of departure from the non‐breeding grounds was more repeatable than timings of arrival at or departure from breeding grounds, suggesting that conditions encountered on migratory journeys and outcome of breeding attempts can influence individual variation.Population‐level shifts in phenology could arise through individual timings changing with environmental conditions and/or through shifts in the numbers of individuals with different timings. Our findings suggest that, in addition to identifying the conditions associated with individual variation in phenology, exploring the causes of between‐individual variation will be key in predicting future rates and directions of changes in migratory timings. We therefore encourage researchers to report the within‐ and between‐ individual variance components underpinning the reported repeatability estimates to aid interpretation of migration behaviour. In addition, the lack of studies in the tropics means that levels of repeatability in less strongly seasonal environments are not yet clear. Repeated tracking of individuals on migratory journeys reveals that there are consistent individual differences in the timing of birds arriving and departing across the annual cycle. Identifying the conditions associated with individual variation in phenology and exploring the causes of between‐individual variation will be key in predicting future rates and directions of changes in migratory timings. Detection of individuals with this method may be incomplete, which may introduce variable lags in observation of the timing of migratory arrivals and/or departures. Lagrangian tracking of individuals through time and space  may therefore be more suited to studies of the timing of individual movements  across the annual cycle, (b) with tracking method, (c) across ecological groups  repeatability metric in our dataset. Second, studies which calculated repeatability using dates when certain latitudes were crossed were excluded unless they were explicitly stated as the arrival or departure dates for the species. We relied on authors' descriptions as to what determines arrival at/departure from the breeding and non‐breeding grounds. Third, we restricted all datasets to breeding adults only. We used this criterion because the refinement of migratory behaviour has shown to be a progressive process mediated by age and experience, particularly for long‐lived species  for this information. One author replied but was unable to provide the requested data, and so neither of these studies was included.To be included in our analyses, observational studies needed to adhere to five main criteria. First, studies had to report repeatability estimates in the form of intraclass correlation coefficients (ICC) using an ANOVA based  from each study, we also extracted the following moderator variables: the annual event for which repeatability was estimated , the method used to track individuals, the coordinates of tagging, and whether this was on the breeding or non‐breeding grounds, study species, sex , the number of individuals (n), the mean number of observations per individual (k) and year of publication. For studies that did not state k but reported the total number of observations, we calculated k by dividing the number of observations by the number of individuals. The methods used to track individuals were grouped into three categories, which represent the type of sampling method (Eulerian or Lagrangian) and the spatial and temporal accuracy of the method: (a) conventional ; (b) geolocation (geolocators); and (c) GPS . If studies used >1 type of tracking method on different groups of individuals, we included both repeatability estimates. Finally, we recorded the statistic that was used to report repeatability (ICC or r), whether any fixed or random effects  were included when calculating repeatability  to the standardised effect size Fisher's Z (Zr) along with the corresponding sampling variance for each study  to zero for our analyses. We used these Zr values and sampling variances (see below) in all meta‐analytical models, but when plotting and reporting parameter estimates we back‐transformed effect sizes to ICC to aid interpretation. The results of all the meta‐analytic and meta‐regression models when including the negative repeatability estimates are reported in the Supporting Information  paper ID, which encompasses multiple effect sizes extracted from the same paper, (b) cohort ID, which encompasses multiple effect sizes obtained from the same group of birds within the same paper, (c) species ID, which encompasses multiple effect sizes from the same species across papers, and (d) effect ID, which is a unit‐level random effect representing residual/within‐study variance. In addition to species ID (a non‐phylogenetic measure), we also included (e) phylogeny (modelled with a phylogenetic relatedness correlation matrix), to account for species similarities due to evolutionary history  which induces a correlation between sampling error variances  method or (b) annual event, due to insufficient sample sizes of certain levels of categorical variables.A multilevel intercept‐only meta‐analytic model was fitted to estimate the overall mean of the effect sizes with the random effects listed above. To evaluate the effects of moderators, we ran a univariate multilevel meta‐regression model for each of the following: (a) tracking method, (b) ecological group, (c) sex, (d) annual event and (e) I2), which indicate the amount of variance unexplained after controlling for sampling variance  as a moderator (a modification of Egger's regression). Likewise, to test for time‐lag bias (i.e. a decline effect), we fitted a multilevel meta‐regression with the year of publication  as a continuous moderator. Finally, we fitted an ‘all‐in’ publication bias test, which included the sampling standard error and year of publication to test for small‐study bias and time‐lag bias, as well as the moderators (above) to account for heterogeneity in our data .A total of 177 effect sizes covering dates of arrival at and departure from breeding and non‐breeding grounds were obtained from 54 papers, including 87 cohorts of birds (Table k of study = 12.4 years). The majority of the extracted repeatability values originated from temperate latitudes in Europe and North America  per effect size were 12 and 39.5, respectively , which is common across ecological meta‐analyses  model revealed a mean repeatability estimate (ICC) for all avian migratory timings across the whole annual cycle of 0.414 , and sample sizes for males and females only were small. None of the other moderators  group , indicating that effect sizes with larger SEs (i.e. more uncertain effect sizes) do not tend to be larger Table , and the4I2[total] > 84%). Repeatability estimates of the four annual events  focused on in this study were found to vary, with departure from the non‐breeding grounds being the most repeatable. However, there was no statistically significant difference in repeatability across ecological groups, the tracking method used to calculate repeatability, between sexes, or with the number of measurements per individual.Advances in tracking technology have allowed the movements of individual birds on repeated journeys to be recorded, which has fuelled interest in the scale of individual variation in migratory journeys. Our meta‐analysis of avian studies tracking the repeat journeys of individuals reveals that repeatability estimates (ICC) of avian migration timing averaged 0.414 (95% CI = 0.3–0.5), although there existed a high heterogeneity (L. l. islandica) to be consistent in spring arrival dates, and that advancing spring arrival dates were driven by new recruits to the population with differing phenology distributions than their predecessors. Changes in the distribution of phenologies within a population could reflect changes in the conditions influencing the development of individual phenologies and/or their subsequent survival rates , there was no statistically significant variation in repeatability values, suggesting consistent individual differences in migratory timings is a common feature of migratory systems  in the sampled population that can be attributed to variation between groups . Therefore, it is important to note that the same repeatability estimates can arise from different patterns of these variance components (see Dochtermann & Royauté, In addition to repeatability in migratory timing, it is also important to consider repeatability in migratory routes and locations. This aspect of migration was not touched upon in this study, but many studies also report high levels of fidelity to breeding and wintering locations (e.g. Delord et al., In conclusion, the similar repeatability estimates of avian migration timing reported by studies of many different species suggest that consistent individual differences in migratory timings is likely to be a common feature of migratory systems. In many cases, repeated collection of individual migration data is not intentional, but rather a by‐product of retrieving a tracking device two or more years post‐deployment. There is also a current gap in the literature with limited information on tropical species, which may limit our understanding of how these species may respond to environmental change in less strongly seasonal environments. As phenological responses to environmental change will depend on the processes that drive within‐ and between‐individual variation and change in migratory timings, methods to disentangle within‐ and between‐individual variation should be incorporated into study designs, for example through structured sampling of individuals across phenological ranges. As migration phenologies are often associated with variation in demographic rates, understanding the consequences of phenological variation will be important for future conservation management strategies and understanding population change.The authors declare no conflict of interest.K.A.F., J.A.G., M.A.C.N., K.N., N.R. and S.J.B. conceived the idea of the study; K.A.F. collected the data, conducted the statistical analyses and wrote the manuscript; S.N. provided statistical advice and support. All authors critically revised the manuscript, contributed to interpreting results and gave the final approval for publication.Appendix S1Click here for additional data file.