ACE1, ACE2, TMPRSS2, IFITM3 and VDR genes remained inconclusive, so we conducted a systematic review and meta-analysis in order to elucidate their potential involvement in the genetic basis underlying the severity of COVID-19 and/or an outcome of SARS-CoV-2 infection. Identification of potentially eligible studies was based on PubMed, Scopus and Web of Science database search. Relevant studies (n=29) with a total number of 8247 SARS-CoV-2-positive participants were included in qualitative synthesis, while results of 21 studies involving 5939 were pooled in meta-analysis. Minor allele I of rs1799752 located within ACE1 was identified as a protective variant against severe COVID-19, while its effect on mortality rate was opposite. Similarly, minor allele A of ACE2 polymorphism, rs2285666, was found to associate with a decreased risk of severe COVID-19 . Statistical significance was also seen for the association between COVID-19 severity and rs12329760 located within TMPRSS2. Our results did not support the supposed association of rs12252 in IFITM3 and polymorphisms within VDR with disease severity. We conclude that genetic variants within ACE1, ACE2 and TMPRSS2 may be potential biomarkers of COVID-19 severity, which needs to be further confirmed in a larger set of studies.Genes involved in the regulation of viral recognition and its entry into a host cell have been identified as candidates for genetic association studies on COVID-19 severity. Published findings on the effects of polymorphisms within  Substantial research efforts have been focused on the discovery of factors predisposing SARS-CoV-2-positive patients to develop a severe disease. Foundations for such investigations were discoveries related to molecular basis of SARS-CoV-2 infection, the immune response in COVID-19, and the observed similarities with other, previously characterized, respiratory infections , a member of the renin-angiotensin system, is the most efficient binding partner . Varianal., 2021; Gómez eal., 2021; Novellial., 2021; Pouladial., 2021). ACE2. Furthermore, a disbalance in ACE1 and ACE2 activities results in vasoconstriction, proliferation, inflammation, tissue injury with fibrosis, thrombotic events and edema, all related to COVID-19 complications . The hypothetical mechanism relies on the activity of this enzyme in the conversion of angiotensin I to angiotensin II, which subsequently affects the expression and function of al., 2020; Guo et al., 2020). A freqal., 2020).The efficiency of viral entry into a host cell is highly dependent on the action of Transmembrane serine protease 2 (TMPRSS2), which cleaves S protein, allowing its fusion with the membrane. In addition, the activity of TMPRSS2 affects the function of Interferon-induced transmembrane protein 3 (IFITM3), which was identified as a potentially relevant factor in the control of SARS-CoV-2 infection, due to its known involvement in other viral diseases, as well as to its interaction with S protein . Since ACE1, ACE2, TMPRSS2, IFITM3 and VDR genes on clinical severity of COVID-19 and/or the outcome of SARS-CoV-2 infection. Since two previous meta-analyses  investial., 2022). In addThe present systematic review and meta-analysis were carried out in accordance with the Preferred Reporting Items for Systematic Review and Meta-Analysis Protocols (PRISMA) .st 2021).The search strategy for potentially relevant studies in electronic literature databases PubMed, Web of Science and Scopus included combinations of keywords: gene/protein name or identifier and aliases ; term “polymorphism” or “genetic variant” and term “COVID-19” or “SARS-CoV-2”. Language restriction was not applied during the search. In order not to miss potentially suitable articles, we thoroughly examined reference lists of the retrieved original articles, as well as previously published reviews and meta-analyses. The selection was limited to articles published in 2020 and 2021  between groups of SARS-CoV-2-positive participants with different COVID-19 severity and/or outcome; b) original full-text articles ; c) sufficient data on genotype/allele counts for the calculation of risk estimates; d) detailed information provided about the recruitment of patients, diagnostic procedure, severity assessment criteria, outcome assessment, ethnicity of participants, genotyping and statistical methodology and other relevant data; e) minimum 50 SARS-CoV-2-positive participants. Studies which provided genotyping data for children and adults separately were included in qualitative and quantitative data synthesis taking into account only results for adults. Eligibility assessment of the retrieved studies was based on the following criteria: a) a study assessed differences in genotype distributions/allele frequencies of polymorphisms located within relevant genes . Separate entries for each study were made for quantitative data synthesis in order to assess all possible comparisons between severity groups, together with fatality rates or other potential outcomes. The data relevant for the extraction from the selected studies included: first author's name and the year of article publication, country and ethnicity of participants, study design, criteria for the diagnosis of COVID-19, severity assessment criteria, patient recruitment method, identifiers of the analyzed genetic variants, genotyping methodology, sample size, age and gender of participants, genotype/allele counts. We reclassified patient groups based on COVID-19 severity in order to match the World Health Organization guidelines, since severity categories differed substantially between studies. Data synthesis was conducted if the results from three or more studies corresponded to a specific comparison of genotype distributions, aiming to evaluate association of a single genetic variant with disease severity/outcome. Methodological quality of studies included in qualitative and quantitative data synthesis was evaluated using Newcastle-Ottawa Scale, which considers three domains  with the highest possible score of 9 . Scorin2), with P value <0.1 or I2≥50 % indicating significant heterogeneity. In cases when the results of heterogeneity tests were insignificant, Mantel-Haenszel method was used for pooling risk estimates under the fixed-effect model  and MetaGenyo , Granada, Spain) were used for assessing between-study heterogeneity and for conducting quantitative data synthesis . Otherwzel, 1959) was useIn cases when the number of entries for a single comparison was more than 5, potential publication bias was evaluated by the visual inspection of funnel plots and by using Egger's test. Robustness of the results of data synthesis was evaluated by sensitivity analysis (leave-one-out method). The study selection process for the present meta-analysis is illustrated by the flow chart given in Figure 1Basic characteristics of the remaining 29 studies . However, not all of them were included in quantitative data synthesis, since one study  investial., 2022; Hubacekal., 2022). ACE2 investigated in more than three studies reporting data which enabled severity/outcome comparison, two studies  which oTMPRSS2, there were 4 eligible studies with enough data to conduct data synthesis of results regarding polymorphism rs12329760. Although Monticelli et al.  did notResults on the association of rs12252 located within IFITM3 with disease severity were extracted from 4 studies. A study of Gómez et al. (2021) was excGenetic variants within VDR were analyzed in 5 eligible studies, which provided enough data for the inclusion in meta-analysis for two polymorphisms, rs2228570 (FokI) and rs731236 (TaqI). ACE1 was 1734, 656 for rs2285666 in ACE2, 2021 for rs12329760 in TMPRSS2, 803 for rs12252 in IFITM3, 1516 for rs2228570 and 1547 for rs731236 in VDR. The overall number of SARS-CoV-2-positive patients from studies included in quantitative synthesis was 5939. Various articles provided data on multiple genetic variants, as well as for multiple groups of patients stratified according to the outcome of SARS-CoV-2 infection, COVID-19 severity and/or lethal outcome, which were used as separate entries in meta-analysis Table 1. The numPrec = 0.039, ORrec= 0.605, 95 % CI = 0.375-0.976; PII vs. DD = 0.047, ORII vs. DD = 0.582, 95 % CI = 0.341-0.993)    in the required number of studies, we performed the analysis for the dominant genetic model . The results suggested that allele A is protective against development of a severe form of COVID-19   . The obtained results suggested that minor allele A confers a protective effect against the progression of COVID-19 to a severe form . In sensitivity analysis for the same group of studies, leaving out the study conducted by Verma et al.  resulteACE1. Even though this polymorphism is intronic, it was shown that the presence of Alu insertion  correlates with the lower expression of ACE1, possibly via regulation of the transcription rate . The pral., 2018; Gómez eal., 2018) demonstal., 2018; Verma eal., 2018). Contraal., 2018), Möhlenal., 2018) and Akbal., 2018) claimedal., 2018).When the presentation of symptoms was analyzed as a criterion for the assessment of the association of rs1799752 with the severity of SARS-CoV-2 infection, Cafiero et al. 2021) reporte reporte2ACE1 and COVID-19 severity for multiple genetic models tested. According to the obtained results, the minor allele I confers a decreased risk of developing severe clinical symptoms. These findings are not surprising, since the majority of studies included in the pooled analysis demonstrated a protective effect of allele I or II genotype, although the authors applied different criteria for the assessment of disease severity. The association with severe COVID-19 was found when patients with severe and moderate illness were compared, most of them being hospitalized for COVID-19. Also, the protective effect of II genotype was confirmed when genotype distributions between patients with mild disease and those with clinically more relevant symptoms were compared. Although statistical significance was not found in the severe vs. non-severe COVID-19 comparison, the obtained results could be explained by differences in the patient recruitment and classification methods.Our meta-analysis demonstrated the association between genetic variant rs1799752 located within Our results are in accordance with those reported in previous meta-analysis by Saengsiwaritt et al. (2022), where Gunal et al. 2021) did not did not2The present meta-analysis showed the association of rs1799752 with a lethal outcome of COVID-19. Since studies included in the analysis varied in their design and in the inclusion of patients with mild disease, we conducted a second round of data synthesis with data only on hospitalized COVID-19 patients. The results of two analytical approaches were similar, but higher OR values were obtained when hospitalized participants were considered alone. However, the minor allele I was found to be associated with an increased susceptibility to lethal outcome, which is in contrast to the results found in a test of its association with disease severity. Explanations for such disagreement may be searched in a relatively low number of participants in the group of deceased patients with COVID-19, in presence of comorbidities that significantly influenced the results of association studies, as well as in differences in transcriptional and posttranscriptional regulation of the expression of ACE1 between a disease and healthy condition. In their study on Greek population, Papadopoulou et al. (2022) found dACE2, Gómez et al.  found nal. (2020) stated al. (2020) failed TMPRSS2 and the risk of severe COVID-19, Monticelli et al.  reporteal. (2021) announcal. . Takingal. . FurtheIFITM3 and a risk of developing severe COVID-19 was found in a relatively large study on Spanish population . Similaal., 2021), althoual., 2021) were stVDR, the meta-analysis was possible for the association of rs2228570 and rs731236 with COVID-19 severity. Apaydin et al. . Still,ACE2 and TMPRSS2 polymorphisms. Further studies are also needed in order to assess the potential influence of ethnicity, as well as to investigate gender differences related to the effect of variants, especially those located on X-chromosome. Additional data would also provide means for testing effects of the examined polymorphisms on specific parameters related to clinical progression of COVID-19, such as presentation of clinical symptoms or COVID-19-related mortality. Future studies should test associations of other genetic variants which were not included in this data synthesis. Finally, it would be highly relevant to test joint effects of all risk variants identified in the present meta-analysis, since genes of interest are functionally related.The results of the present meta-analysis clearly qualify rs1799752 for a genetic variant associated with COVID-19 severity. Since the number of studies included in this meta-analysis of the effects of all polymorphisms tested is relatively low, a caution is suggested when considering the obtained results. Another limitation of our meta-analysis is a small number of patients in certain severity- or outcome-based groups. Therefore, additional studies with larger sample size are needed to confirm the significance of the associations found in the current data synthesis, including those on the effects of The research was supported by the Ministry of Education, Science and Technological Development of the Republic of Serbia (Agreement no. 451-03-68/2020-14/200019).ZD: Conceptualization, investigation, formal analysis, visualization, writing - original draft. DR: Investigation, formal analysis, visualization, writing - review & editing. NG: Investigation, formal analysis, visualization, writing - review & editing. MŠ: Investigation, formal analysis, writing - review & editing. AP: Investigation, formal analysis, writing - review & editing. GM: Investigation, formal analysis, writing - review & editing. ON: Validation, supervision, project administration, funding acquisition, writing - review & editing. The authors declare that they have no conflict of interest.The data supporting the findings of this study are available within the article, its supplementary materials, or from the corresponding author, upon reasonable request.


Q4 = 1.77, 95% CI = 1.32–2.38), cholesterol , and palmitoleic acid , and an inverse association with high intake of dietary heptanoic acid  and low intake of palmitic acid . None of the fat variables were associated with rectal cancer. Our study suggests that the recommendation of limited consumption of fats may decrease the risk of CRC among the Iranian population.The evolving trends in colorectal cancer (CRC) as one of the most common malignancies worldwide, have likely been influenced by the implementation of screening programs and changes in lifestyle habits. Changing lifestyle, including the shift in diet composition with higher fat, sugar, and animal-source foods intake, led to an increasing burden of CRC in countries undergoing rapid socioeconomic improvement. Results for the link between specific fatty acids (FAs) and CRC are generally inconclusive and more limited in developing countries than elsewhere. This study aims to investigate the association between FA intakes and CRC and its anatomical subsites in a large Iranian case-control study. A food frequency questionnaire was used to collect information on dietary intake in 865 cases and 3206 controls. We conducted multivariate logistic regression models to calculate the odds ratio (OR) and 95% confidence interval (CI). We found positive association between CRC and high intake of dietary total fat (OR highest quartile  Colorectal cancer (CRC) is the third most common malignancy and the second leading cause of cancer death, with an estimated 1.8 million new cases and 900,000 deaths in 2020 worldwide . The chaAt least 12% of CRC cases are directly attributable to overweight/obesity according to a recent global review . AccordiCurrently, screening programs in Iran are based on the identification of high-risk individuals, corresponding to first-degree relatives of CRC patients. Thus, they are based on the consultation of cancer registry. The first-degree families of CRC patients are invited to cancer screening and participate in counseling sessions, possibly leading to colonoscopy recommendation. So far, Iran does not have any screenings or plans for screening the general population at medium-risk of cancer. This makes Iran very different from most of the other countries, where non-invasive testing is purposed based on risk profile but also older age and are followed by the invasive endoscopic examination. This issue was addressed by Nikbakht et al. who conducted a study to investigate the use of immunochemical fecal occult blood test (IFOBT) in a mid-risk for CRC Iranian population. The study showed high responsiveness rates from the population, and high rates of positive IFOBT were found . ConsideThis study aims to look at associations between fatty acid intakes and CRC risk considering overall and sub-site-specific CRC in a large Iranian case-control study. Our analysis provides useful data to deepen the knowledge on the role of fatty acids in CRC development. Moreover, we give valuable information on the consumption of fatty products in Iran and their relationship with CRC from different anatomical sites.n = 18); those with missing information on dietary intake and those in the highest or lowest 1% of the distribution for the ratio of energy intake to estimated energy requirement (n = 60). We included 865 cases and 3,206 controls in our analysis . They were part of the IROPICAN study, a multicenter case-control study coducted in 10 Iranian provinces, which was designed to examine the link between opium use and the development of lung, colorectal, bladder, and head and neck cancers . For theanalysis . Controlanalysis , assuminCRC cases was defined by the International Classification of Diseases (ICD-O-3) as tumors of the colon (C18) or rectum (C19–C20). Colon cancer may be categorized as proximal  and distal , while rectum cancer occurs from the recto-sigmoid junction (C19) down to the rectum. Anus tumors were excluded from the study. Also, all histological types of CRC except melanoma and sarcoma were included.During a face-to-face interview, lifestyle questionnaires were used to collect information about education, tobacco use, opium use, socioeconomic status (SES), physical activity, previous illness, and use of nonsteroidal anti-inflammatory drugs (NSAIDs). At the time of enrollment, trained professionals generally measured the standing height of participants (cm). The body mass index (BMI) was calculated as weight/height squared (kg/m2). Cases were asked about their body weight before cancer diagnosis, while controls were measured their body weight at the time of the interview. Based on the Finland Job Exposure Matrix (FINJEM) , 24 we aDietary intakes were assessed with validated qualitatively Persian Cohort FFQ  administRegarding this analysis, we calculated the intake of 50 dietary fats based on our food composition table. According to previous studies , 22, 30 The means and standard deviations (±SD) for continuous variables and the frequencies for categorical variables were calculated for descriptive statistics of baseline characteristics and intake of dietary factors of cases and controls.The normality tested by comparing a histogram of the data to a normal probability curve. After examining the distribution of the data, all nutrient intakes were log-transformed to improve normality.p values <0.05 as statistically significant.For the analysis of the association between dietary fat and different type of FFAs; SFAs, PUFAs, MUFAs, and cholesterol and CRC, we conducted unconditional multivariable logistic regression models to estimate the odds ratios (ORs) and 95% CIs. Before multivariate logistic regression, correlations between different exposures were checked and ORs were adjusted by gender , age (continuous), province, BMI (continuous), tobacco (smoking and water pipe) consumption (Never /Ever), aspirin use (Yes/No), opium use (no user / irregular users / regular users'), SES (Low/ Medium/ High), work-related physical activity (sedentary/Moderate/Heavy), processed meat intake including mortadella, hamburger and sausage , calcium , fiber intake , and energy . Participants with missing data for physical activity (24.93 %) were coded as distinct categories during the analysis. Quartile were calculated based on the distribution of different type of fat intake between controls of study. Besides the main analysis, continuous analyses were also run across quartiles. Furthermore, analyses were stratified by age (under and over 50), BMI, gender, SES, physical activity, and vegetable intake for comparing quartile 4 to quartile1 of different types of fat intake. Nevertheless, according to significant interaction (P-heterogeneity < 0.05) analysis we only report age category results. We repeated the analyses for all CRC patients and then after stratification for sub-sites, including colon overall, proximal colon, distal colon, and rectum. All statistical analyses were carried out using Stata 14 . We considered The study was approved by the Ethics Committee of the National Institute for Medical Research Development (NIMAD) (Code: IR.NIMAD.REC.1394.027). All participants signed a written informed consent to participate in the study.N = 145) of the colon cancers were in the proximal colon, 42% (N = 185) in the distal colon, and 23% (N = 104) to the overlapping region between proximal and distal colon. In our study, the male to female ratio was 2, and around 76% of all participants were older than 50 years of age. Fars and Tehran provinces had the largest share of the study population. Significant differences between cases and controls were observed for tobacco use, family history, socioeconomic status, and opium use, but not for BMI and physical activity. CRC cases reported higher dietary energy intake  compared to controls . Higher levels of FAs and cholesterol were reported by cases than controls , cholesterol , and palmitoleic acid , as well as an inverse association with high intake of dietary heptanoic acid , low intake (quartile 2) of palmitic acid   of dietary total fat (mean31–0.88) .Q4vs.Q1 = 1.43, 95% CI; 1.15–1.77], and distal colon .Additional analyses were performed on specific subsites of CRC cases. CRC were classified as colon, proximal or distal colon cancer, and rectum cancer. Based on stratified analyses by anatomical site, we identified a positive association between total fat and colon cancer  as well as proximal  and proximal colon cancer risk .Also, cholesterol intake was positively associated with colon , myristic acid  and proximal colon cancer. Also, an inverse association was observed for heptanoic acid and colon cancer . Intake of oleic acid, which is one of the important MUFAs, was inversely associated with colon  and proximal colon cancer risk . Conversely, no significant associations were observed for total MUFAs and PUFAs. Rectal cancer was not associated with any FA  and the source of the fat , 30–32. in-vivo and in-vitro studies  pro-inflammatory effects that may be triggered by interaction between dietary fat and gut microbiota, which play a role in the metabolism of bile acids (BA) , 38. In  studies , 43. (4) studies . (5) Mor studies . A doubl studies .Our results showed IROPICAN study population reported a higher daily intake of total fat , cholesterol 253 mg/day), and saturated fat  than recommendations, coupled with a lower intake of total PUFAs , reflecting previous studies regarding the amount of dietary fat from different sources in Iranian populations , 48. The mg/day, Over time, the type of oil available on the Iranian market has changed. According to a WHO report published in 2018, Iranian households consume a great deal of cooking oil made from partially hydrogenated vegetable oils, a major source of trans fatty acids (TFAs) . In receTo our knowledge, this is the first study to investigate the magnitude of different types of FAs on CRC in a large Iranian population from different provinces and in the EMRO region. Our data are characterized by high quality because (i) information were collected by trained interviewers, (ii) same standardized validated FFQs and questioning tools were used in the different centers, and (iii) all cases were provided with pathological confirmation and allowed us analyses by subsites.Two potential limitations of our study include: (i) selection bias, because the controls were not chosen on a population-based approach in the primary study (IROPICAN) but were rather taken among the healthy visitors who did not have CRC or other diseases. However, our validation study showed that due to appropriate using healthy visitors instead of disease controls such bias is minimal ; (ii) reIn this large CRC case-control study, total fat, cholesterol, and higher intake of myristic acid, palmitic acid, palmitoleic acid from animal sources were associated with increased risk of CRC, and some FAs from plant sources such as heptanoic acid or oleic acid- the main FA in olive oil- decreased particularly colon cancer risk, after accounting for major adjustments. Moreover, subgroup analyses by age revealed that participants older than 50 years had a higher risk of CRC due to consumption of a high FAs diet. This may be due to the cohort effect and changes on the amount, and type of FAs intake over time. Our study improves general knowledge on CRC epidemiology and offers important insights on CRC in Iranian population. These data may be useful for the identification of high-risk individuals and public awareness to promote prevention of CRC in Iran and other LMICs that facing the increasing pattern in the incidence of CRC. In line with international evidence, the promotion of a decrease in fat consumption, especially FAs from industrial and animal sources, may decrease the risk of CRC among the Iranian population. Along these lines, a number of restrictions have been imposed in Iran since 1025 on oil products, including trans and saturated fat, aimed at controlling and preventing non-communicable diseases, including CRC .The raw data supporting the conclusions of this article will be made available by the authors, without undue reservation.The study was approved by the Ethics Committee of the National Institute for Medical Research Development (NIMAD) (Code: IR.NIMAD.REC.1394.027). Written informed consent was obtained from the individual(s) for the publication of any potentially identifiable images or data included in this article.MSS, KZ, and PB designed research. MSS, KZ, RA-N, RS-F, HR, and MH conducted research. MSS, GC, KZ, and PB analyzed data, performed statistical analysis, and wrote the paper. FK, AE, MG, VC, IH, and EP contributed in the editing of the preliminary results and draft. KZ had primary responsibility for final content. All authors read and approved the final manuscript.This study was funded by National Institute of Medical Research Development (NIMAD) (Code: IR.NIMAD.REC.1394.027) and particularly supported by Investigator Grant No. 24706 of Fondazione AIRC. Italy.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.Where authors are identified as personnel of the International Agency for Research on Cancer / World Health Organization, the authors alone are responsible for the views expressed in this article and they do not necessarily represent the decisions, policy or views of the International Agency for Research on Cancer / World Health Organization.


The prognostic significance of insulin‐like growth factor binding protein 2 (IGFBP2) expression has been explored in plenty of studies in human cancers. Because of the controversial results, the meta‐analysis was carried out to evaluate the relevance of IGFBP2 expression with the prognosis in various tumors.The data searched from four databases  was used to calculate pooled hazard ratios (HRs) in this meta‐analysis. Subgroup analyses were stratified by ethnicity, cancer type, publication year, Newcastle–Ottawa Scale score, treatments, and populations.Twenty‐one studies containing 5560 patients finally met inclusion criteria. IGFBP2 expression was associated with lower overall survival  and progression‐free survival  in cancer patients, but not with disease‐free survival  or recurrence‐free survival . The subgroup analyses indicated IGFBP2 overexpression was significantly correlated with overall survival in Asian patients , Caucasian patients , glioma , and colorectal cancer  and surgery subgroups .The meta‐analysis showed that IGFBP2 expression was associated with worse prognosis in several tumors, and may serve as a potential prognostic biomarker in cancer patients. The prognostic significance of Insulin‐like growth factor binding protein 2 (IGFBP2) expression has been explored in plenty of studies in human cancers. However, the results remain controversial. Here, we conducted a systematic review and meta‐analysis to assess the relevance of IGFBP2 expression with the prognosis in various tumors. It was estimated that there were 18.1 million new cancer cases and 9.6 million new cancer deaths in 2018.Studies have demonstrated that the IGFBP family (IGFBP1–6) are relevant in many cancers.IGFBP2 was found to be highly expressed in many cancers, including glioma,Therefore, we performed the meta‐analysis to assess the association between IGFBP2 expression and the survival outcomes and the prognostic significance of the IGFBP2 in cancer patients, searched from 21 literature studies containing 5560 patients.2The study was carried out according PRISMA Guidelines.2.1A comprehensive and accurate search of articles was performed by Pubmed, Embase, Cochrane Library, and Web of Science up to February, 2021. The search strategy is shown in 2.2Inclusion criteria included studies that: (1) explored the connection between IGFBP2 expression level and prognosis in various cancer patients; (2) provided sufficient data by table or KM curve; and (3) were published in English. Exclusion criteria included: (1) duplicated publications; (2) animal studies, cell line experiments, reviews, letters or case reports; and (3) without available or useable data.2.3The data collection and quality evaluation were completed by two researchers independently. Any disagreement would be discussed and solved by group discussion. The following data were extracted from eligible studies by a standardized information collection: first author's name, the time of publication, author's country, the gender of patient, the age of patient, the clinical stage, the type of cancer, the method of detection, outcome, study quality, and follow‐up months. Because of different definition methods in included studies and expression levels in various cancers, IGFBP2 overexpression in this study was defined according to individual definitions from the included studies. The quality of each study was evaluated by the Newcastle–Ottawa Scale (NOS). At the same time, if the NOS score was ≥7, the study would be recognized as high quality.2.4I2 statistics was used to assess the power of heterogeneity. An I2 value of 0% means no observed heterogeneity, with larger I2 values meaning enhanced heterogeneity. The pooled HR was calculated by random‐effect model when the p‐value was <0.05 and I2 was >50%; otherwise, a fixed‐effect model was used. Subgroup analyses stratified by cancer type, ethnicity, NOS score, publication year, treatments, and populations were carried out to further evaluate the general sources of heterogeneity. Besides, sensitivity analysis was conducted to assess the constancy of each result in this meta‐analysis. Moreover, if publication bias was detected, the number of supposedly unpublished studies to adjust for publication bias was evaluated with a trim and fill method.Analyses of all data were completed using Stata 12.0 software. The strength of the connection between IGFBP2 expression and prognostic items  of the cancer patients was evaluated with the HRs and the corresponding 95% CI. In addition, the Higgin's 33.1We could understand the process of study selection from Figure The primary feathers of all articles included in this study are summarized in Table 3.2Data for the association between IGFBP2 expression and various cancer prognosis are shown in Figures 3.3To explore the source of heterogeneity of OS, subgroup analyses of OS was performed. The major results of this subgroup analyses are shown in Table 3.4P < 0.0001). The number of unpublished study assessment was 7 using the trim and fill method as shown in Figure P < 0.0001).The stability of the results was evaluated through sensitivity analysis pooled HRs for OS by eliminating each study, respectively. No significant changes were reported for the pooled HRs. Therefore, the result of the meta‐analysis was proved to be credible Figure . At the 4In this meta‐analysis, we conducted the quantitative analysis to evaluate the relationship between IGFBP2 expression and prognosis in various cancer patients. Besides, we performed subgroup analyzes based on ethnicity, cancer type, publication year, NOS score, treatments, and populations. The HR values for OS and PFS were 1.57 (95% CI = 1.31–1.88) and 1.18 (95% CI = 1.04–1.34), respectively, which showed that there was a significantly negative correlation between IGFBP2 expression and OS, PFS in cancer patients. This connection was also observed in both Asians and Caucasians. When analysis was restricted to glioma and colorectal cancer, we found a worse prognosis. Although IGFBP2 expression was related to poor PFS, the result should be interpreted cautiously because of limited included studies. In addition, although IGFBP2 expression was not connected with DFS and RFS, all three articles included in this meta‐analysis for DFS suggested there was a significant connection between IGFBP2 expression and DFS. This may be due to the limited number of articles. Therefore, these results need to be confirmed by more studies.IGFBP2 is found to be highly expression in many of cancers,We also explored the potential source of the high heterogeneity based on PRISMA Guidelines. The subgroup analyses in accordance with ethnicity, publication year, NOS score, treatments, cancer type, and populations were performed. There was still highly significant heterogeneity among 11 studies of Asians, 5 studies of Caucasians, populations >100 and < 100 subgroups. Besides, we found that the heterogeneity decrease in neither before 2016 nor 2016–2021 subgroups. Therefore, ethnicity, publication year, and populations were not the main source of the heterogeneity for OR in this meta‐analysis included 16 studies. We found that the high heterogeneity among included studies reduced obviously according to the treatments. That is to say, treatments were likely to be a source of high heterogeneity. When the analysis was limited to the caner type, we found that the heterogeneity of colorectal cancer decreased obviously. In the same way, the heterogeneity of NOS score <7 also reduced significantly. The heterogeneity could be explained through cancer type and NOS score. The additional source of heterogeneity might be associated with the method of HR and 95% CI obtained from KM curve. We gained HR and 95% CI by Engauge Digitizer and Excel provided by Tierney directly when these data were not reported by the authors.Several studies related to fields indicated high IGFBP2 expression is correlated with worse prognosis for glioma patients.The strength of our study was to systematically collect evidence on the prognostic value of IGFBP2 in cancer patients. This makes it possible to use a risk prediction model of including IGFBP2 to forecast prognosis of cancer patients. This meta‐analysis was restricted to the following points. First, some studies only provided KM curve without direct HR and 95% CI, which might influence the result's accuracy. Second, there was significantly high heterogeneity among included studies. Third, this was a retrospective analysis of published studies. As a result, it is vulnerable to these published studies and contributes to several publication bias. Besides, the positive results which tend to be easier to publish is one of the important factors in publication bias, which leads to exaggerating the relationship between IGFBP2 expression and OS.5To sum up, the present meta‐analysis shows that there is a significant association between high IGFBP2 expression and worse OS and PFS in various cancer patients. Thus, IGFBP2 may serve as a great biomarker of prognosis in various cancers, and it is worthy of further study to add it to the established risk prediction model to provide clinical the basis for individualized treatment plan. Nevertheless, data are insufficient to determine its role cautiously. More prospective studies of large number of patients are needed to further confirm the value of IGFBP2 in various cancers.YWX and YHP contributed to the study design; BZ and CQH searched the literature, BZ and LFW analyzed the data and wrote the manuscript; YHL, YL, and CQH drew the figures and improved the language. All authors read, revised, supervised, and approved the final manuscript.This work was supported by funding from the Natural Science Foundation of China (81972801), the Basic and Applied Basic Research Foundation of Guangdong (2019A1515011873), the Li Ka Shing Cross‐Disciplinary Research Grant Foundation (2020LKSFG01B), the Science Medical Project and Shantou Technology Planning (200605115266724), and the Science and Technology Program of Guangdong Esophageal Cancer Institute(Q201906).The authors declared that there is no conflict of interest for this article.Ethical consent was not required for the systematic review and meta‐analysis as all data obtained from previously published studies.Appendix S1 Supplementary InformationClick here for additional data file.


Autoimmune thyroid diseases (AITDs) and antiphospholipid syndrome (APS) are commonly seen in childbearing-age women and can lead to recurrent miscarriages. Studies with a relatively small sample size were conducted and concluded inconsistent results on the relationship between AITDs and APS or the presence of antiphospholipid (aPL) antibodies. This meta-analysis aimed to provide evidence on the possible association between AITDs and APL using primary data from all relevant studies.  Literature databases including PubMed, Embase, and Cochrane were searched from the time when the databases were established to March 2022. A total of 7 studies that met all inclusion criteria were selected in this meta-analysis, with 475 AITD patients and 764 healthy controls. After data extraction, the pooled odds ratio (OR) and the 95% confidence interval (95% CI) were calculated. p < 0.001). There was no significant association between the risk of Graves' disease (GD) and positive APL . The risk of positive APL in patients with Hashimoto's thyroiditis (HT) increased over 2 folds compared with healthy controls .  The risk of positive APL in AITD patients was approximately 3 folds higher in AITD patients compared with those in healthy controls (OR 3.41, 95% CI 2.29–5.08,  The result of this meta-analysis proved that the presence of APL was significantly increased in patients with AITD compared with healthy individuals, especially in patients with Hashimoto's thyroiditis. Autoimmune thyroid disorders (AITDs), including Graves' disease (GD) and Hashimoto's thyroiditis (HT), are one of the most common organ-specific autoimmune diseases. HT manifests as clinical hypothyroidism due to thyroid autoimmunity (TAI), in other words, the autoantibodies against thyroglobulin and/or thyroid peroxidase. It was reported that HT was a T helper (Th) cell-mediated autoimmune disease, which resulted in diffuse lymphocytic infiltration of the thyroid gland . The maiAntiphospholipid syndrome (APS) is an autoimmune disease characterized by recurrent thrombosis and miscarriage due to the presence of antiphospholipid (aPL) antibodies. APS consists of primary conditions or secondary to other underlying autoimmune diseases, most commonly systemic lupus erythematosus (SLE). A series of studies have proved the existed association between AITD and SLE . In a meA literature search of PubMed, Embase, and Cochrane was conducted for studies from the time when the databases were established to March 2022 with the restriction of English language using the following searching terms: “antiphospholipid antibody” or “antiphospholipid syndrome” in combination with the terms “thyroid diseases” or “Hashimoto's thyroiditis” or “Graves' disease” or “autoimmune thyroid disease.” The studies were filtered for those focused on human subjects. All eligible studies and their eligible references were checked and retrieved. Studies were selected by two independent reviewers, and disagreements were solved after discussion.p value.Studies were included in this meta-analysis when they met the following criteria: (1) case-control or cohort design; (2) evaluate the association between AITD and patients with positive aPL; (3) containing sufficient data of cases, healthy controls for calculating the odds ratios (OR), 95% confidence interval (CI), and the The following data were extracted from all enrolled studies independently by two of the authors: first author, publication year, numbers of cases of AITD and healthy controls, and numbers of cases and controls with positive aPL.The Newcastle–Ottawa Scale, recommended by the Cochrane collaboration for risk of bias assessment for observational studies, was applied for all included studies for quality assessments by two independent investigators . Three cp < 0.05 was considered statistically significant. Heterogeneity was evaluated by chi-squared statistics and I2. Values of 25%, 50%, and 75% were defined as low, moderate, and high heterogeneity, respectively. The random-effects model was used when I2 >50%, indicating heterogeneity across studies. Otherwise, the fixed-effects model was used in the meta-analysis. Funnel plots were performed for analysis of publication bias. All statistical analyses were performed using review manager software (version 5.3).The association between AITD and positive aPL was investigated by OR with 95% CIs. A total of 239 articles were observed during the preliminary search. Among them, 171 articles were excluded after the authors screened the titles and abstracts. The rest of the 68 publications were identified and read in the full article, as shown in p < 0.001), which suggested the risk of positive aPL was approximately 3 folds higher in AITD patients compared with those in healthy controls  (p=0.33)  , which ip=0.002)  , which dA sensitivity analysis was performed using sequential omission of the included studies to evaluate the reliability and stability of the conclusions. The pooled ORs were not significantly changed after the analysis.Funnel plots were conducted to evaluate the potential publication bias for meta-analysis . The disTo the best of our knowledge, our study is the first systemic review and meta-analysis aiming to explore the association between AITD and the presence of APL. In our study, we discovered the positive association between AITD and positive aPL, as well as HT and the presence of aPL.AITD, as well as aPS, is common in women of reproductive age that can cause recurrent pregnancy loss and infertility . There aThe positivity of aPL antibodies was not explored in patients with AITD, nor was the prevalence of antithyroid antibodies detected in patients with APS. APS could be a primary clinical condition or secondary to other autoimmune diseases including SLE. The positivity rate of antithyroid antibodies in SLE patients has been widely discussed, while the association of both diseases highly varied in the literature –19. DiffOf all 7 studies included in the meta-analysis, only 1 study evaluated the prevalence of confirmed diagnosed APS in AITD patients and healthy controls in accordance with the classification criteria for definite APS . Therefoβ2-glycoprotein I antibody (antiβ2-GP I). aPL profile was divided into medium-high titers, high-risk, and low-risk based on different risks for thrombotic and obstetric APS [β2-glycoprotein I antibody, thus we failed to conduct further subgroup analysis for different subtypes of aPL or risk for thrombotic and obstetric APS.aPL included lupus anticoagulant (LAC), anticardiolipin antibodies (ACL), and antitric APS . A high-Our study is the first meta-analysis to shed light on the association between AITD and the presence of aPL, while there were several limitations to the study. First, due to the lack of consistent diagnostic criteria for APS among different studies, the study population included patients with confirmed diagnosed APS and recurrent miscarriage patients with positive aPL antibodies, which increased the heterogeneity of the meta-analysis. Second, the included studies were conducted by researchers from different countries for the duration of up to three decades, and the diversity of the included studies may lead to possible influence on the study results. Third, several studies with a large study population of positive aPL or AITD were not enrolled in this meta-analysis due to the lack of the comparison of healthy controls or incomplete raw data, thus the study population of our meta-analysis was relatively small. Therefore, the results of this meta-analysis should be interpreted with caution.In conclusion, the result of this meta-analysis proved that the prevalence of positive aPL was significantly increased in patients with AITD compared with healthy individuals, especially in patients with Hashimoto's thyroiditis. More population-based studies, especially from Asia and other continents, are expected in the future.


Oat beta-glucan is reported to have a beneficial effect on lowering blood lipids. To probe the effect of oat beta-glucan consumption on serum lipid profiles , we carried out a systematic search on randomized controlled trials of oat beta-glucan intervention on hypercholesterolemic individuals. (2) Methods: the pieces of literature were obtained from PubMed, Scopus, Cochrane Library, Web of Science, and the Embase from inception to 28 February 2022. The results were presented with the weighted mean difference (WMD) with a 95% CI. The random-effects or fixed-effects model was applied according to the heterogeneity. The subgroup analysis and meta-regression were used to identify the source of heterogeneity. (3) Results: thirteen trials with 927 participants were included in our meta-analysis. Overall, oat beta-glucan supplementation significantly reduced levels of TC , LDL-c . Furthermore, beta-glucan consumption did not show significant effects on TG , HDL-c . Subgroup analysis indicated that critical factors, such as disease severity of participants, the daily intervention of oat beta-glucan, source of oat beta-glucan, and duration of intervention had impacts on outcomes. (4) Conclusions: oat beta-glucan intake may significantly decrease the level of TC and LDL-c while no significant changes in TG and HDL-c were observed. This meta-analysis supports the health benefits of oat beta-glucan, especially for its cholesterol-lowering features, although it has some inevitable limitations. Cardiovascular diseases (CVD) have been an important cause of human death for decades. According to the report from the American Heart Association, the crude prevalence of CVD was 485.6 million cases in 2017, and this number increased by 28.5% over the past 10 years ,4. AlthoOats, as one of the most cultivated and consumed whole grains worldwide, have been increasingly appreciated after adjustment to the dietary structure. With the progression of nutrition, oat was recognized as a healthy food in the mid-1980s signifying that a substance in it helped prevent heart disease and therefore it became popular for human nutrition ,11. SimiAlthough Zhu et al. conducted a quantitative assessment to evaluate the effects of beta-glucan on lipid profiles in hypercholesterolemic subjects in 2015 [We conducted our search according to The Preferred Reporting Items for Systematic Reviews and Meta-Analysis guidelines . The seaStudies were selected if they met the following inclusion criteria: (1) the design of study was a randomized controlled trial; (2) participants were patients aged over 18 years old with hypercholesterolemia and had no restrictions on complications; (3) trials using oat beta-glucan as an intervention to evaluate its effect on blood lipid profiles; (4) participants in the control group took a placebo or appropriate diet; (5) both the intervention group and control group had clear indicator values or numerical changes in blood lipid profiles.The exclusion criteria were as follows: (1) objects involved in studies were animals or cells; (2) participants took drugs or other substances at the same time which might affect the level of the blood lipid profile; (3) trials without accurate data including total cholesterol (TC), total triglyceride (TG), high-density lipoprotein-cholesterol (HDL-c), and low-density lipoprotein-cholesterol (LDL-c) from baseline to endpoint; (4) blended interventions.Two reviewers (J.Y. and J.X.) extracted the following characteristics of 13 trials independently: first author, year, study region, study design, sample size , age, sources of β-glucan, amounts of β-glucan, duration, comparison group, baseline and endpoint of different lipid levels. The results of data extraction were aggregated after rigorous discussion.Quality assessment of the studies included in this meta-analysis was performed according to the Cochrane collaboration risk-of-bias tool . Seven dnet change = q1 for the 1st quartile, M for the median, q3 for the 3rd quartile), where n represents the sample size, and p < 0.05. Heterogeneity between trials was assessed by I2 and chi-squared statistics, either I2 > 50% or a p-value < 0.05 was considered as significant heterogeneity. In addition, a fixed-effects model was adopted to conduct the meta-analysis if statistical heterogeneity was present (p > 0.05 or I2 < 50%); otherwise, a random-effects model was used. Subgroup analyses were performed to explore the sources of heterogeneity. To identify which trial(s) led to the heterogeneity, a leave-one-out sensitivity analysis was adopted. In addition, we conducted subgroup analyses and regression based on the characteristics of disease severity, daily intervention dose, source of β-glucan, and duration. All statistical analyses above were performed using Stata version 15 .We included trials with oat beta-glucan intervention or added oat beta-glucan to certain foods or the diet. Under the premise of meeting the previous inclusion and exclusion criteria, we split the trials into high and low doses and used “LD” to represent low doses and “HD” to represent high doses. For studies that did not provide direct data, we calculated the mean net change in blood lipid profiles by subtracting the mean change . In addition, we calculated the mean difference (MD) in parallel-controlled trials by using the change from baseline in the control and intervention groups. If there were several endpoints in a study, we extracted the last value as the calculation. As for cross-over trials, the end of lipid profile levels in the control group and intervention group were selected to compute the MD in the effect. Data presented in milligrams per deciliter were transformed to millimoles per liter by dividing by 38.66 for HDL-c, LDL-c, and TC while 88.6 for TG. The formula was used if the report did not provide the net change of SD: SDs et al. . MeanwhiStudy selection and screening were documented and are presented in Overall, there are 13 RCTs summarized in Thirteen studies investigated the effects of supplementing oat beta-glucan on TC, TG; ten studies probed the effects of oat beta-glucan consumption on HDL-c while twelve trials explored the effects of oat beta-glucan supplementation on LDL-c in patients with hypercholesterolemia.p = 0.000), and there exists no significant heterogeneity between all the trials .Four trials reported that the intervention of oat beta-glucan had no significant effects on TC while ten remaining trials indicated that oat beta-glucan could lower the level of total cholesterol. However, when all the studies above were involved in our meta-analysis, the result revealed that consumption of oat beta-glucan did cause a significant reduction in TC in patients with hypercholesterolemia . In addition, significant heterogeneity was exhibited between included trials .Two studies found that oat beta-glucan significantly decreased TG while two other trials reported that oat beta-glucan increased total triglyceride level. However, when all the data from included studies were pooled for meta-analysis, it revealed that consuming oat beta-glucan had no significant reduction in TG in patients with hypercholesterolemia . In addition, heterogeneity between trials was significant .Reyna-Villasmil et al. reported that oat beta-glucan could remarkably increase HDL-c while Queenan et al. and Kerckhoffs et al. found that the oat beta-glucan group significantly decreased the level of HDL-c compared with the control. Moreover, when all the data from trials were included in the analysis, the results showed that oat beta-glucan intake had no significant effects on HDL-c in patients with hypercholesterolemia , and there was significant heterogeneity between the included trials .Eight included studies demonstrated that oat beta-glucan significantly decreased LDL-c while other trials did not. A cumulative meta-analysis of thirteen RCTs indicated a significant reduction in LDL-c after consuming oat beta-glucan in comparison with placebo-control groups in patients with hypercholesterolemia (mean difference = −0.27; 95% CI: −0.35 to −0.20; More detailed results for the effect of oat beta-glucan on lipid profiles are presented in The results of subgroup analysis are presented in p-value of 0.374 suggested that the results of TG are reliable. As for the other three indicators, visual scanning of funnel plots and Egger’s regression p-value of 0.637, 0.366, and 0.519 indicated an absence of publication bias, respectively. The funnel plots are depicted in Potential publication bias was measured by funnel plots and Egger’s regression test. We evaluated the symmetry of the funnel plots intuitively. Besides, a method of “trim and fill” analysis was conducted to evaluate publication bias further if there existed any asymmetry in the funnel plot . AlthougIt is well known that an elevated lipid profile is associated with the pathogenesis of cardiovascular diseases. Consuming more than 3 g oat beta-glucan each day is recommended by the FDA to gain its beneficial effects. This meta-analysis of 13 RCTs appraising the oat beta-glucan intake on lipid profiles indicated that oat beta-glucan intake has significant impacts on TC and LDL-c but not on TG and HDL-c. Despite the health benefits of consuming oat beta-glucan, few meta-analyses were found to evaluate its single effect on hypercholesterolemic individuals. Xu et al. conducted a meta-analysis focusing on the different effects of several delivering matrices of oat beta-glucan in mildly hypercholesterolemic subjects which concluded that a combination of “liquid” and “solid” products exerted the most remarkable effect . AnotherBased on the current subgroup analyses, we found several factors that were worth further exploration. As detailed above, only two studies targeting moderate hypercholesterolemic subjects were considered for inclusion. According to the degree of disease, we found the different effects of oat beta-glucan on lipid profiles. When we conducted a sensitivity analysis, we found that four major indicators of lipid profiles almost decreased significantly in these two trials compared with control groups while other trials targeting mild hypercholesterolemic subjects had controversial results ,27. ThisUnlike general intervention, the effects of β-glucan could also be explained from a microscopic standpoint. Kerckhoffs et al. held the opinion that the effect was decreased owing to the bile acid reabsorption and different molecular weight β-glucans . The conExtrusion and processing probably improve the bioactivity of oat β-glucan . Oat β-gBeyond the effects of oat beta-glucan intake on lipid profiles in hypercholesterolemic adults, it was reported that oat β-glucan mixed with various vitamins and minerals in oat noodles could significantly reduce the level of both TC and LDL-c. In addition, a great decrease in many CVDs ratio markers was also witnessed in this trial . MoreoveThere are several strengths in our meta-analysis. First, an examination of β-glucan′s single effect made it more powerful than any mixed intervention trial. Second, we comprehensively assessed the effect of oat β-glucan on lipid profiles in hypercholesterolemic adults; this provided a systematic view of elucidating its bioactive activity mechanism. Meanwhile, this meta-analysis focused on the gender differences and different severity levels of disease for the first time. We hope these findings based on current evidence will be a helpful contribution to a deeper exploration of oat β-glucan.Simultaneously, the present review has several apparent limitations. First, we failed to find the precise sources of heterogeneity in TG, HDL-c, and LDL-c via subgroup analysis and meta-regression. This might be caused by other factors that we ignored. Second, owing to the careful screening of the including studies, trials using oats or oat bran that did not report a clear amount of oat β-glucan were omitted. The results of these trials may potentially affect our conclusion. Thirdly, this meta-analysis was not persuasive enough given its limited sample size.Overall, the results from our analysis demonstrated that dietary consumption of oat beta-glucan may significantly reduce the level of TC and LDL-c in patients with hypercholesterolemia. Meanwhile, there is insufficient evidence to determine whether or not an association exists between oat beta-glucan intake and changes in TG and HDL-c levels. Therefore, we recommend that hypercholesterolemia patients take in adequate oat beta-glucan to help manage lipid profiles. Besides, more large-scale, high-quality randomized controlled trials are needed to support these findings since the current trials had inconsistent results.


Sirtuins (SIRTs) have key roles in cancer progression. However, the prognostic implications of SIRTs in breast cancer (BC) remains a subject of debate and controversy. Thus, we performed a meta-analysis to identify the precise prognostic value of SIRTs in BC patients.Systematic literature searching was conducted in PubMed, Cochrane Library, Web of Science, and Embase databases. The pooled hazard ratios (HRs) with 95% confidence intervals (CIs) were calculated to estimate the association of SIRTs expression and survival outcomes in BC patients.A total of 22 original studies with 6317 patients were eligible for this meta-analysis. The results showed that in patients with BC, elevated SIRTs levels were associated with shorter overall survival (OS) and disease-free survival (DFS) both in univariate  and multivariate analysis models . Notably, further subgroup analysis revealed that overexpression of SIRT1 and SIRT6 predicted poor OS  and DFS  in BC.Our data has elucidated that SIRT1 and SIRT6 could serve as prognostic biomarkers for patients with BC and may contribute to refined patient management.The online version contains supplementary material available at 10.1007/s12672-022-00529-7. Breast cancer (BC), the most common malignancy worldwide, remains the pivotal cause of cancer-related mortalities for women . On the +)-dependent enzymes [Sirtuins (SIRTs), which share homology with the yeast silent information regulator 2 (Sir2) gene, are a family of highly conserved nicotinamide adenine dinucleotide  AND  AND (“SIRT1” OR “sirtuin 1” OR “SIRT2” OR “sirtuin 2” OR “SIRT3” OR “sirtuin 3” OR “SIRT4” OR “sirtuin 4” OR “SIRT5” OR “sirtuin 5” OR “SIRT6” OR “sirtuin 6” OR “SIRT7” OR “sirtuin 7”). All the references from the selected articles were furtherly scanned to identify more eligible studies.The selection of studies was completed independently by two authors (HZ and CM). The inclusion criteria were as follows: (a) having overall survival (OS), disease-free survival (DFS) or recurrence-free survival (RFS) outcomes; (b) patients were divided into a high expression group and a low expression group; (c) the diagnosis of BC was histologically and pathologically confirmed; (d) availability of sufficient data to calculate the hazard ratio (HR) and 95% confidence interval (CI); and (e) publication in English. The exclusion criteria were as follows: (a) duplicate or overlapping populations; (b) reviews, meta-analyses, case reports, letters, animal studies, and conference abstract; (c) studies without survival outcomes; and (d) studies published in a non-English language. Disagreements between the two reviewers were discussed to reach a consensus.Data extraction was done independently by two investigators (HZ and RH), including first author, year of publication, original nation, molecular subtype of BC, the studied SIRT type, sample size, antibody-related information, detection method, cut off value, survival outcome, HR with its 95% CI and analysis model. If a study did not report the HR, the survival information from Kaplan–Meier curves were digitized and extracted using Enguage Digitizer 4.1 . Any conQuality assessment was independently conducted by the same two researchers (HZ and RH) according to the Reporting Recommendations for Tumor Marker Prognostic Studies (REMARK) guidelines . The REM2 test, with greater than 50% considered as high heterogeneity. A random-effects model was employed for analyses in the presence of substantial heterogeneity, otherwise, the fixed-effects model would be applied. Considering the almost identical definition, the DFS and RFS outcomes were merged in the meta-analysis. Subsequently, all the univariate and multivariate HRs were combined to evaluate the effect on prognosis respectively. To find the potential source of heterogeneity, the sensitivity analysis was performed. In addition, potential publication bias was detected using the funnel plot and Begg’s test. P value less than 0.05 was considered significant.All the statistical analyses were done with the software STATA 16.0 . Heterogeneity among studies was determined using the chi-square-based Q and IThe initial search yielded a total of 750 potentially eligible records from the online databases. After removing duplicates, we screened 615 studies for further assessment. Then, 442 records were excluded based on studying the titles and abstracts after application of the exclusion criteria and 173 selected for full-text screening. Ultimately, 22 publications comprising 6317 samples were enrolled in this meta-analysis –39. The The 22 eligible articles were published between 2011 and 2020, from 7 different countries and the sample sizes ranged from 48 to 688. The expression level of SIRTs was mainly measured by immunohistochemistry (IHC), while only 4 studies used quantitative real-time polymerase chain reaction (qRT-PCR) as the detection method , 32, 34.I2 = 85.9%, p = 0.000) and multivariate  analysis models, the random-effects model was selected to compute the pooled HR and its 95% CI. Regardless of the SIRTs type, we found that higher SIRTs expression significantly linked with poor OS both in univariate and multivariate analyses  analysis. As for the multivariate model, a slight asymmetry in the funnel plot was shown to the right side of the pooled estimates .The association between SIRTs expression and OS in BC was evaluated in 20 studies in a total of 5217 patients –39. In tI2 = 74.1%, p = 0.000; I2 = 69.4%, p = 0.003; respectively). Similarly, both the pooled results of univariate and multivariate estimates indicated a significant association between elevated SIRT6 (nuclear) expression and poor OS in patients with BC . Additionally, stratified analysis by different molecular subtypes of BC revealed that elevated SIRT1 expression predicted a significantly worse OS in TNBC patients  through multivariate analysis. However, no statistically correlation was noticed between SIRT3 expression and OS neither in TNBC  nor in HRBC  patients. Detailed results are shown in Table To explicate the heterogeneity, subgroup analysis was further conducted based on the type of SIRTs Table . As showI2 = 74.9%, p = 0.000) and multivariate analysis . As shown in Fig. p = 0.063; p = 0.293; respectively) did not suggest the existence of obvious publication bias  and multivariate analysis . Meanwhile, higher levels of SIRT6 (nulcear) expression were statistically associated with poorer DFS  in fixed-effects model through univariate analysis .Similar to OS analysis, the subgroup analysis was performed according to the type of SIRTs for DFS. As shown in Fig. The association between SIRT family members and survival outcomes of BC patients has been widely investigated, but no compatible results have been achieved. Numerous studies have discovered that SIRT1 overexpression was correlated with unfavorable OS in BC patients –20. HoweHistone deacetylases (HDACs) are major agents of epigenetic regulation and their dysfunctional deacetylase activity has been strictly related to the tumorigenesis process . SIRTs (+/+ mice, in vivo dada showed that Sirt6 deletion repressed mammary tumor development and increased survival in Sirt6+/− mice [Being a predominantly nuclear member of the SIRT family, SIRT6 has a pivotal role in DNA repair, genome maintenance and glucose metabolism . Many st+/− mice . Nuclear+/− mice . Besides+/− mice . Given t+/− mice . HoweverSIRT3 is capable of metabolic reprogramming and contributes greatly in the fate of cancers. Up to date, tumor-suppressive and oncogenic roles of SIRT3 were both discussed in BC . The mosNevertheless, this analysis has several limitations. First, only studies published in English were included in our meta-analysis, which is likely to result in selection bias. A bias may also exist due to the positive results reported in most of the including publications, which may have led to overestimations in our analysis. Second, there was statistical heterogeneity in this analysis, which might be partially explained by different antibody sources, different cut-off values and different detection methods. Third, some HR values were extracted from Kaplan–Meier curves, which may partly compromise the precision of the data. Forth, a difference in protocols for treatment after surgery in various studies might have a great impact on survival outcomes and thus resulted in some heterogeneity. Finally, the small sample sized articles lacking statistical power may affect the reliability of the results.In conclusion, our meta-analysis highlights that elevated SIRTs levels were significantly associated with shorter OS and DFS via combining univariate and multivariate analyses. Particularly, SIRT1 and SIRT6 could serve as prognostic biomarkers for patients with BC and may lead to refined patient management. However, these results should be validated in well-designed prospective cohort studies, and the clinical role of SIRTs for BC deserves further investigation.Additional file 1. Raw data.


Neural tube defects (NTDs) are congenital abnormalities in the central nervous system. The exact etiology of NTDs is still not determined, but several genetic and epigenetic factors have been studied. Folate supplementation during gestation is recommended to reduce the risk of NTDs. In this review we examine single nucleotide polymorphisms (SNPs) of the genes in the folate pathway associated with NTD. We reviewed the literature for all papers discussing both NTDs and SNPs in the folate pathway. Data were represented through five different genetic models. Quality assessment was performed using the Newcastle–Ottawa Scale (NOS) and Cohen’s Kappa inter-rater coefficient assessed author agreement. Fifty-nine papers were included. SNPs in MTHFR, MTRR, RFC genes were found to be highly associated with NTD risk. NOS showed that high quality papers were selected, and Kappa Q-test was 0.86. Our combined results support the notion that SNPs significantly influence NTDs across the population, particularly in Asian ethnicity. Additional high-quality research from diverse ethnicities is needed and meta-regression analysis based on a range of criteria may provide a more complete understanding of the role of folate metabolism in NTDs. Folates are part of the B-group family of vitamins consisting of an aromatic pteridine ring bound to p-aminobenzoic acid and a glutamate residue through a methylene bond. Folate is an essential vitamin; it cannot be synthesized by human metabolism and needs to be derived entirely from the diet. Folic acid is the synthetic source of folate found in fortified foods and many vitamin supplements . FolatesUpon ingestion, folate is bound to folate-binding protein (FBP). Folyl poly-γ-glutamate carboxypeptidase (FGCP), which is encoded by the folate hydrolase 1 (FOLH1) gene, is an apical brush border transmembrane enzyme found in the small intestine which releases folate from the bonded proteins in the intestinal lumen. Two folate carriers located on the apical surface of the enterocyte, the proton-coupled folate transporter (PCFT) and the reduced folate carrier (RFC), transport folate from the intestinal lumen into the enterocytes . The mulIn addition to the RFC and PCFT transmembrane folate carriers, there are several receptor mediated folate transporters. Specifically, three glycoproteins known as folate receptor alpha (FRα), beta (FRβ) and gamma (FRγ) mediate cellular folate uptake .The human body’s metabolic systems use the reduced form of folate. Folic acid, the synthetic form of folate, is an oxidized form of folate. Oxidized folates, like folic acid, need to be reduced to tetrahydrofolate (THF) through dihydrofolate reductase (DHF). THF is the entry into the folate cycle. The folate cycle supports two important metabolic pathways, the production of purines and the methylation cycle. Through the folate cycle, THF is converted to 5,10-methylene–THF by the pyridoxal phosphate dependent enzyme serine hydroxymethyltransferase (SHMT) using L-serine and producing glycine in the process. 5,10-methylene–THF interconverts to 5-formyltetrahydrofolate, commonly known as leucovorin or folinic acid, which is the precursor for purine nucleotides. 5,10-methylene–THF is converted into 5-methyltetrahydrofolate (5-methyl-THF) through the enzyme methylenetetrahydrofolate reductase (MTHFR). 5-methyl-THF is the folate form that drives the methylation cycle. Methionine synthase (MTR), a vitamin B12-dependent enzyme, converts 5-methyl-THF back to THF to complete the folate cycle, and in the process produces methionine from homocysteine (HCY).Methyl donors are essential for regulation of gene expression and enzyme activity. Methionine produces S-adenosyl methionine (SAM), the body’s major methyl donor, by methionine adenosyltransferase (MAT) through an adenosine triphosphate (ATP) dependent reaction. Donating a methyl group results in the production of S-adenosylhomocysteine (SAH). SAH is then converted in a reversible reaction to HCY and adenosine by S-adenosylhomocysteine hydrolase (SAHH) through a reversible reaction. In addition to the reaction using 5-methyl-THF, methionine can also be produced from HCY using betaine-homocysteine methyltransferase (BHMT), an enzyme exclusively found in the liver and the kidney. BHMT converts betaine and homocysteine to dimethylglycine and methionine. Importantly, the methylation cycle is connected to transsulfuration reactions which produce glutathione, the major antioxidant of the body. HCY produces cysteine, the rate limiting precursor to GSH production, through two pyridoxal phosphate-dependent reactions catalyzed by cystathionine β-synthase (CBS) and γ-cystathionase (CTH).Neurulation, the process of closure of the neural tube, occurs 21 and 28 days after conception. Interruption of this process can cause neural tube defects (NTDs), resulting in congenital deformities of the brain and spinal cord . NTDs inFolate was found to be severely insufficient in the diets of women who had offspring with NTDs, according to ground-breaking research by Smithells et al. . FurtherTo participate in the methylation cycle and contribute to the metabolism of folate, folic acid must first transform into the naturally bioactive form known as THF. A deficiency in neural tube development may be brought on by inhibition in the folate metabolism pathway. The genes in the folate pathway are being extensively studied for this reason, and it has been shown that proteins encoded by certain important genes, including MTHFR, RFC, and methionine synthase reductase (MTRR), work together with MTR to transfer the methyl group to homocysteine ,17. WhenIt’s interesting that despite numerous research studies examining the relationship between NTDs and SNPs, no consensus has been achieved. Folate metabolism may be inhibited by folate pathway gene SNPs ,25,26,27To ensure the rigor of this meta-analysis, we followed the Preferred Reporting Items for Systematic Reviews and Meta-Analysis (PRISMA) guidelines .We searched three databases, PubMed, EMBase and Web for published papers, from June 1996 to June 2022, which examined any SNPs relating to the folate pathway associated with NTDs. The search strategy was based on English keywords, “MTHFR”, “MTRR”, “RFC”, “folate pathway” “polymorphism” or “SNP” and “NTDs or Neural Tube Defects or Spina Bifida”. References for the articles and the collected studies were reviewed.The following inclusion criteria had to be met: (1) case–control research and cohort study design; (2) data on some, any, or all SNPs in the folate pathway; (3) presentation of data appropriate for the measurement of odds ratios (ORs); (4) consistent description of NTDs. Animal research, maternal studies, case reports and studies without original data were omitted. Duplicated studies were removed.All data were analyzed independently by two reviewers (A.K.A. & R.E.F.). The following information was extracted from the qualifying literature: year of publication, first author’s name, country, ethnicity, genotyping process, source of control, and case matching control variables. Allele or genotype counts were extracted or estimated from the published data for the re-calculation of crude ORs and their 95% confidence intervals (95% CIs) for the assessment of the association of SNPs in the folate pathway with NTDs.p < 0.05 was regarded as statistically significant. All five genetic models  were used to compute the aggregated estimate of the OR and corresponding 95% CI for the association. Egger’s exact test was used to assess heterogenicity between papers. The pooled ORs and 95% CIs were calculated using a fixed-effects model (the Mantel-Haenszel technique) when there was no substantial heterogeneity (p > 0.1 or I2 50%) between trials; otherwise, a random-effects model (the DerSimonian-Laird method) was chosen. We conducted subgroup analyses by ethnicity to examine the impact of between-study heterogeneity on our pooled data. By using Pearson’s chi-square test, Hardy–Weinberg equilibrium (HWE) in controls was assessed, with p < 0.05 indicating a departure from HWE. By excluding each selected study one at a time and reanalyzing the remaining papers, sensitivity analysis was used to determine the stability of the meta-analysis results. Additionally, the sensitivity analysis was carried out by removing studies that violated the HWE. The Egger’s linear regression test and the Egger’s quantitative test were used to assess any potential publication bias. A significant publication bias was deemed to exist when the Egger’s test plot was asymmetric, and the p-value for the Egger’s test was less than 0.05.Only SNPs with 4 or more reports that included at least 5 homozygous cases for the risk allele were included in the analysis. ORs with 95% CIs were used to evaluate the strength of the link between the SNP and the risk of NTDs. The pooled OR values were determined using the Z-test, and Analysis of the quality of the final selected papers was evaluated using the Newcastle Ottawa Scale . The scaThe selection and identification of literature are shown in p ≤ 0.0001), recessive , homozygote , heterozygote , dominant  (p > 0.10) and the funnel plot shows no heterogeneity  . Egger’sogeneity .p > 0.10) but funnel plots showed heterogeneity  (p = 0.75) and the funnel plot showed no heterogeneity  . The Eggogeneity .p = 0.01) (p = 0.73) but the funnel plot show heterogeneity  . The Eggogeneity .p = 0.49) and funnel plots showed little heterogeneity for the dominate model  models. Funnel plot showed no heterogeneity and Egger’s test show no bias for the allele model  for Asian and Caucasian ethnicities in 3 of the 5 genetic models. Only Asian ethnicity was significant in the recessive and heterozygous models for MTHFR. The RFC association with NTDs was only seen in the Asian ethnicity and it was significant in 4 of the 5 models. None of the models were significant for Caucasian ethnicity.Subgroup analysis was conducted using all 5 genetic models to examine differences between ethnicities and SNPs in the SNPs where associations were found in the overall analysis see . Ethnic Newcastle-Ottowa Scale was conducted to assess quality of the literature by 2 authors (A.L.L. & A.R.W.). Results were represented as a scale of 1–10. Majority of the papers selected had a score of 7 and above which represents good quality . Cohen’sNTDs are the most prevalent congenital disorder, second only to cardiovascular disorders . The effThis study identifies MTHFR C677T, MTRR A66G and RFC A80G SNPs as risk factors for NTDs. Our findings are summarized in Birth abnormalities can result from gene mutations in essential enzymes that control folate metabolism and transportation. Several mutations can severely affect MTHFR activity, reducing folate status and potentially explaining a quarter of the frequency of NTDs ,73. TT mLoss-of-function mutations in MTR and mutations of the MTR chaperone MTRR may influence homocysteine levels resulting in severe disease phenotypes ,77. Our The association of RCF SNPs with NTDs is still unclear. Relton et al.  found thMTHFD1, also known as “C1-THF synthase,” is a trifunctional cytoplasmic enzyme that is nicotinamide adenine dinucleotide phosphate (NADP) dependent. Additionally, the functional SNP MTHFD1 G1958A has been investigated in several populations, and earlier research has shown that MTHFD1 genetic variations can cause a variety of disorders ,84,85,86A good technique for combining data from all relevant research to increase statistical power is meta-analysis. However, a few concerns with this meta-analysis need to be considered. First, the number of mixed population studies that were considered for subgroup analysis was limited, with the bulk of the subjects being Asians. Additionally, following stratification, the sample sizes for Caucasian ethnicity was reduced, which may have limited the statistical power of our findings. Second, we were unable to locate any articles written in languages other English and Chinese, which could have influenced our results. Third, some studies with unfavorable findings tend not to be published , which could bias the available data. Fourth, NTDs encompass a wide spectrum of malformations, including encephalocele and spinal bifida. Additionally, the severity and prognosis of NTDs vary along a continuum, from anencephalic stillbirth to meningocele spina bifida with mild symptoms. Therefore, if studies only include live births and less severe instances, the effects of SNPs on the development of NTD types may be understated. Finally, epigenetic variables may be linked to the occurrence of NTDs. Due to the lack of data related to gene–gene and gene–environment interactions in the formation of NTDs these potential contributors to NTDs could not be studied.Despite these caveats we believe our meta-analysis is robust in its findings. First, sensitivity analysis showed that none of the included studies independently changed the meta-analysis results. Indeed, removing the HWE-violating studies had no effect on the outcomes of our pooled data. This suggests that the pooled results are sufficiently reliable and stable. Second, this is the most thorough meta-analysis to include data from earlier research on the connection of the SNPs in the folate pathway with risk of NTDs and spina bifida. In conclusion, our combined findings are consistent with the idea that SNPs are significantly associated with NTDs across the population, notably in the Asian ethnicity.The current study, which performed a cumulative meta-analysis of 42 studies of MTHFR C677T that provided results with a relatively narrow 95% confidence range, significantly supports the association between the MTHFR C677T SNPs and NTD risk. We found a possible link between RFC A80G and the risk of NTDs but this was only confirmed for Asian ethnicity, potentially because of the contradictory results in studies performed in the Caucasian population. Significant results were found for the MTRR SNP but for only one model. In contrast, we were unable to identify an association for the remaining SNPs. Future work will benefit from the incorporation of additional high-quality studies from various ethnicities when they become available. Additionally, when further studies start to incorporate more environmental variables, meta-regression analysis based on a variety of additional parameters of interest may help reduce heterogenicity between studies.Most importantly, we have come to appreciate the differences between folic acid, which is an oxidized folate compound with limited bioavailability. Indeed, we are understanding that reduced folate compounds such as folinic acid, leucovorin and methyltetrahydrofolate may be able to bypass blockage in the folate pathway in order to prevent and treat disease . Thus, a


AGT) gene to be significantly associated with BD across all 7 studies. We detect a commonly significant BD-specific subnetwork with the ESR1 gene as the main hub of a subnetwork, associated with neurotrophin signaling (p = 4e−14) and myometrial relaxation and contraction (p = 3e−08) pathways. ancMETA provides a new contribution to post-GWAS methodologies and holds promise for comprehensively examining interactions between genes underlying the pathogenesis of genetic diseases and also underlying ethnic differences.Over the past decades, advanced high-throughput technologies have continuously contributed to genome-wide association studies (GWASs). GWAS meta-analysis has been increasingly adopted, has cross-ancestry replicability, and has power to illuminate the genetic architecture of complex traits, informing about the reliability of estimation effects and their variability across human ancestries. However, detecting genetic variants that have low disease risk still poses a challenge. Designing a meta-analysis approach that combines the effect of various SNPs within genes or genes within pathways from multiple independent population GWASs may be helpful in identifying associations with small effect sizes and increasing the association power. Here, we proposed ancMETA, a Bayesian graph-based framework, to perform the gene/pathway-specific meta-analysis by combining the effect size of multiple SNPs within genes, and genes within subnetwork/pathways across multiple independent population GWASs to deconvolute the interactions between genes underlying the pathogenesis of complex diseases across human populations. We assessed the proposed framework on simulated datasets, and the results show that the proposed model holds promise for increasing statistical power for meta-analysis of genetic variants underlying the pathogenesis of complex diseases. To illustrate the proposed meta-analysis framework, we leverage seven different European bipolar disorder (BD) cohorts, and we identify variants in the angiotensinogen ( The main goals of trait mapping studies, including genome-wide association studies (GWASs), are to understand the genetic architecture of diseases, pinpoint the number of loci associated with a particular trait, and approximate the underlying heritability rate . Once thin silico replication to maintain reliability in genetics association findings These limitations reflect a gap in our understanding of the mechanisms underlying the pathogenesis of complex traits and diseases. The major source of these shortcomings is the method of GWAS itself as restricted to a single-marker-based testing approach  cohorts, and we finally outlined the implications, challenges, and opportunities that cross-ancestry meta-analyses present in the GWAS era. The proposed method has been implemented in the ancMETA tool Here, we discuss the proposed meta-analysis framework, ancMETA. It performs meta-analysis at two different levels by aggregating multiple independent population GWAS summary statistics datasets. It uses an integrative analysis through Bayesian posterior probability and combines the results into known biological protein–protein network datasets. Lastly, ancMETA performs the meta-analysis at the subnetwork level and identifies the most significant subnetworks to understand the biological pathways .We describe six different steps of the proposed meta-analysis framework as follows:N-independent studies.Step 1: Collection of N ≥ 2 independent GWAS summary statistics datasets, from the same phenotype or trait. GWAS summary statistics are defined here as per-genetic locus effect sizes (log odds ratios) together with their standard errors, p-values, or z-scores for the affected–unaffected traits This step requires Step 2: Mapping SNPs to the associated genes.This is an intermediate step, and it is similar to our previous approach ith study  has Ji genes N is the total number of studies. Let jth gene of the study i. For an SNP p-value, respectively. The proposed meta-analysis framework assumes that the SNPs related to one gene are correlated. This assumption is due to the fact that each gene contains a large number of SNPs mostly under LD; therefore, the magnitude of the statistical outcomes for each SNP within a gene from each study is considered to be approximately the same. Thus, the effect size and the p-value will follow a normal distribution. Let jth gene i . Thus, a fixed-effects model can be applied to estimate the combined effect size at the gene level from each study i = 1, 2, … N. We construct a linear function of parameters We assume that the Let μ ≠ 0. Although the solution, which corresponds to the maximum true effect size Note orStep 3: Meta-analysis at the gene level.Let Gj is μj is an unknown parameter. The same calculations as in step 2 are applied to estimate the effect size of the gene Gj across the studies, which is calculated as follows:where and its standard error is given as follows:ρi of the study ith for i = 1, 2,…, N is equal to Q) is given as follows:τ2 = 0. If we consider the expectation of If the precision df), then for Q < df (which means τ2 < 0), the maximum between 0 and τ2 is considered, so that τ2 is non-negative. Recalling that in step 2, we pointed out that SNPs within one gene are correlated. Therefore, to perform meta-analysis, a fixed-effects model is used Since the test statistics in Step 4: Mapping genes to biological networks.For this step, all the genes and their related statistical outcomes and gene–gene combined LD are mapped into a PPI network database , which cμ is the unknown true effect size at the subnetwork level with K genes, the estimated effect size is given by Step 5: Meta-analysis at the subnetwork level. This step estimates the effect size and standard error of each subnetwork from . μ ≠ 0, maximizing the log likelihood by solvingSince yieldsStep 6: Computing the overall statistical significance.N studies, and let us apply the symmetrical property of the Gaussian distribution. Therefore, the posterior probability is approximately given asFrom step 5, we know that τ2 is the between-study variance. Recalling that p-values can be evaluated through a sampling method developed by However, a test statistic based on the weighted sum of chi-square can be used, under the null hypothesis that there is no association. Thus, from the fact that The U.S. residents of northern and western European ancestry (CEU), Yoruba (YRI), and Mexican (MEX) populations from the HapMap3 Project were used to generate independent case–control studies. Details of these populations can be found in We conducted a GWAS on each dataset using EMMAX CBX7, LD0C1L, and ASTN1 associated with our simulated causal SNPs , of which AGT has a strong effect across the European populations including Irish, Scottish, Norwegian, and British in contrast to the European American  GWAS summary statistics obtained from the NIMH data repository . These dAmerican .ESR1) gene is the main hub. ESR1 interacts with AGT and is connected with the other gene hubs with known BD-associated genes such as CACNA1C, PLCG2, NCAN, BCR, and BDNF. e−14) and other interesting biological pathways such as myometrial relaxation and contraction (p = 3e−08), morphine addiction (p = 9.2e−08), DNA damage response (p = 3e−05), and alcoholism (p = 2e−05). In addition, the subnetwork (p = 7.7e−21) and is also associated with autosomal dominant (p = 2.8e−07) inheritance types of diseases (Leveraging the recent version of the human PPI network (IntAct release 239) from the IntAct database bnetwork  is implidiseases .AGT, CACNA1C, ESR1, NCAN, BDNF, BCR, and GSK3B as well as with the other protein-coding genes in To evaluate the determined interconnectivity among Designing a post-GWAS meta-analysis that leverages the combined effect of multiple SNPs within a gene or genes within subnetworks/pathways across multiple GWAS datasets may reveal consensus association signals and identify large effect sizes. This may further provide valuable information in prioritizing the most important results across different populations. Here, we proposed a Bayesian graph-based gene- and pathway-specific meta-analysis approach. We implemented the proposed model in ancMETA , which aA well-reconstructed human protein–protein interaction network is a powerful tool in network biology and medicine research, which forms the basis for multi-omics and dynamic analyses In addition to the GWAS summary statistics, ancMETA allows users to use any weighted biological network and accepts a user-defined network. In case if users provide an unweighted biological network, ancMETA leverages case–control genotype datasets to construct the weights of the network , in whicSome improvements need to be considered in future work, such as accurately modeling the convergence of the SNP signal to the related subnetworks and leveraging the weakly/moderately associated signals from different GWAS studies. It is worth mentioning that we have applied ancMETA on old BD GWAS with a limited sample size see . HoweverThere is also a need to integrate summary-level data across multiple phenotypes to simultaneously capture the evidence of the aggregate-level pleiotropic association. The lack of accurate knowledge of complex traits and the sensitivity of human protein interaction network makes it challenging to directly compare the results from the different pathway analysis methods. Overall, the method implemented in the proposed framework highlights the value of identifying the effect size of pathways associated with a disease, which may be useful in understanding the pathogenesis, disease risk prediction, and susceptibility to genetic diseases.CACNA1C and NCAN have previously been implicated in BD through GWAS AGT has not reached GWAS significance in the previous studies despite being considered as a candidate in BD due to its role in the renin–angiotensin system AGT from this analysis (p = 3.2e−19) therefore strengthens its association with BD and further highlights the potential impact of ancMETA as a useful tool in discovering additional small effect variants that may be missed in the single GWAS. Importantly, the very interesting AGT finding is due to the innovations made in addressing population variation and to the inclusion of a PPI framework. The results from the subnetwork analysis revealed a strong interaction of the hub gene ESR1 with AGT, as well as CACNA1C, NCAN, BCR, GSK3B, and BDNF, suggesting the fact that the AGT finding here is not merely indicative of a false positive, but rather that is valid. In addition to a moderate link to BD, ESR1 has been related to migraine onset, alcohol dependence, obsessive compulsive disorder, and postpartum depression The results obtained from the GWAS summary statistics on the European BD cohorts found association with seven genes, of which