Breast cancer is the world's most prevalent malignancy, with an increasing incidence and a predisposition for postpubertal females from all cultural and ethnic backgrounds. More recently, oral Fusobacterium nucleatum species have been observed in cancerous human breast tissue, drawing attention to the role of microbes in cancer pathogenesis.  Investigating oral Fusobacterium nucleatum species as potential biomarkers for female-specific breast cancer. st January, 1983–31st March, 2022. Articles included were in English and based on women between the ages of 18–96 years with confirmed gingivitis/periodontal disease and breast cancer diagnoses from registered specialists. Authors extracted data independently, and a meta-analysis of risk estimations measuring associations between oral Fusobacterium nucleatum species and female-specific breast cancer was elucidated via calculated relative risks and 95% confidence intervals.  A systematic search in The Central Register of Controlled Trials, EMBASE, EBSCO, NCBI, and MEDLINE databases was undertaken from the 1I2 = 41.39%; P = 0.02), and the importance of periodontal status on breast cancer pathogenesis was determined .  AXIS tool analysis revealed 78.70% of articles with a positive correlation between oral Fusobacterium nucleatum and female-specific breast cancer. The risk of breast cancer development increased with significant levels of oral Fusobacterium nucleatum due to gingivitis/periodontitis . Low-moderate statistical heterogeneity was found ( Oral Fusobacterium nucleatum species are a risk factor for breast cancer development, thus elevating their biomarker potentiality. Breast cancer (BC) is the world's most prevalent malignancy, with a predisposition for postpubertal females . In 2020Despite all international efforts to raise public awareness, outreach programmes, frequent testing for the condition in women above 40 years, and extensive research into the various cell-signalling pathways responsible for BC development, its incidence is rising amongst postpubertal females of all cultural and ethnic backgrounds –3.Attention has been drawn to the significance of microbes and their role in cancer pathology initiation/activation, with further support for an association between disruptive microbial homeostasis and BC –8. The oWhen the intraoral environment becomes overburdened with dental plaque accumulations, this can cause immediate microbial dysbiosis, initiating acute inflammatory responses that result in gingivitis . AccumulThe results highlight the actively prominent role played by oral FN species and endothelin molecules in the creation of a perfect proinflammatory microenvironment evident in individuals diagnosed with PD and its link to accelerating BC pathogenesis –25.Over time, innovative periodontal surgical techniques have come into play with the purpose of slowing/arresting PD progression and regenerating lost periodontal tissues. Several flap surgical designs have been proposed with modifications to minimise periodontal surgical trauma (PST); however, the traditional method to access flap surgery  remains the advocated surgical procedure for pocket reduction/elimination . This isMIS is a novel approach that aims to respect hard and soft tissues via minimal flap reflections, thus reducing postoperative pain, encouraging healing, and improving clinical outcomes . Since 2The significance and functionality of the human microbiome are mainly derived from a plethora of studies into the human gut, the origin of approximately 99% of the microbial mass responsible for the maintenance of homeostatic gross metabolic function . ComparaThis systematic review was created in accordance with the preferred reporting system for systematic reviews and meta-analyses (PRISMA) . Also reThe P.I.C.O.S model of clinical questioning for evidence-based medicine was implemented by FG, RCG, and RGM as the defining criteria for article incorporation, utilising the P.I.C.O.S point system , EMBASE, EBSCO, NCBI, and MEDLINE. The time period from which databases were searched by FG, RCG, and RGM was from February 1criteria .In total, three separate searches were conducted and categorised into 3 phases: phase 1 was the initial search undertaken by FG, RCG, and RGM with the purpose of collating and evaluating articles linking human microbiome dysbiosis to cancer development. Phase 2, undertaken by FG and RCG, was a more “concise” search of articles focussing on a dysbiotic relationship between varying ecosystems belonging only to the human oral microbiome and cancer development. The final stage, supervised by FIG (phase 3), focussed on articles investigating a link between oral FN bacterial species and their influence on female-specific BC pathogenesis.In all 3 phases, abstracts and titles were independently reviewed by FG and RCG. FG collated the required data from the selected articles to be incorporated, and data extraction were cross-checked by FG, RCG, and RGM. In instances where author discrepancy was discovered in phase 1, articles were reanalysed and progressed to phase 2. If discrepancies remained, a consensus was reached by requesting involvement by the co-authors, with FG having the final say.The methodological sections of 6 out of the 10 articles were analysed by FG, RCG, and RGM utilising the AXIS Tool  were considered for inclusion. Our focus for this systematic review was on cancer development and not on survival/mortality rates.To examine specific bacterial communities, DNA microarrays create high-throughput, quantitative and systematic analysis—especially when examining oral microbiota . HoweverFrom the 10 articles included, 6 articles underwent AXIS Tool and meta-critical analysis –43 and 4n = 49), 189 relevant articles were included (79.41%).Initially, 238 articles, including duplicates, were retrieved. 164 studies were excluded (68.91%), and following duplicate eliminations (25 articles were eligible for inclusion (13.23%), with a final total of 10 studies eventually included for analysis (40.0%), consisting of 1745 subjects . Articlen samples = 45,622, n patients = 1348), 3 in South America , 2 Europe . The female population age range was 18–96 years, with a generalised trend of case participants being older than healthy female control participants. An extended range of articles were sourced, with a range of publication dates from 1983–2022.Our results revealed that 5 studies were implemented in North America  [n = 1) [n = 4) [A more detailed evaluation of the data provided by the 6 included articles was implemented, and details of article study design, follow-up periods, and dental status were examined . How artn = 1) , case-co [n = 1) , and pro [n = 4) –40, 43.I2 = 41.39%; P = 0.02; The findings highlighted a statistical difference between the included studies  . FurtherWith respect to the evaluation of specified follow-up durational data from the included articles, it was apparent that 2 studies , 39, whiThe result of this systematic review reveals a link between oral FN species and their role in female-specific BC pathology, thus elevating their potential to act as biomarkers in BC. However, from our subgroup calculations, it can be elucidated that numerous cofounders, such as follow-up periods and study designs, may have influenced the overall evaluations.Over the past decade, much research effort has been instigated with particular emphasis to the understanding of oral microbial burden impaction on an individual's overall systemic health, and numerous articles –48 have The Human Oral Microbiome Database (HOMD) was conceived and setup in 2005, providing vital information on the types of bacterial niches present within the oral cavity . Within Although the establishment of polymicrobial films as essential participants in the initiation and progression of periodontitis has been validated, some questions remain unanswered—the most important being how microbial dysbiosis and fluctuating microbial compositions within the oral cavity are able to instigate periodontal pathogenesis and if specific microbial characteristics relate to other systemic diseases. More recently, the emergence of meta-transcriptomics and metagenomics has permitted high-throughput analysis of microbial communities, thus illuminating some aspects of this conundrum.Traditionally, microbial detection and identification involved the implementation of culture methodologies, including sample preparations, plate counts, and the isolation of singular bacterial colonies . HoweverIn 2022, a review article examined the possible mechanisms by which oral microbial FN species were able to colonise the lactiferous ducts of breast tissue, via translocation into the systemic circulatory system . A furthFap2 of FN species is a galactose-inhibitable adhesin involved in coaggregation and cell adhesion . ResultsA comparative study in 2016 undertook extensive analysis of microbial communities originating from female breast tissue samples of both benign and malignant growths. The findings deduced that there was a direct correlation between the microbial enrichment of FN species and that of female-specific BC malignancies . The infContradictorily, a study in 2003 evaluating data from 11, 328 participants with a follow-up period between 11 and 21 years, including the NHANES I follow-up within the aforementioned time period, demonstrated no relationship between gingivitis or periodontitis and BC diagnosis in adult women . A furthAn imperative aspect to consider is the importance of PD surgical procedures, the influence of “flap designs” and utilisation of dental loupes on the risk of developing periodontitis. Conventionally, DFAs are the primary procedures implemented for direct root access and pocket elimination. Nonetheless, there are innumerable disadvantages such as membrane exposure, flap dehiscence, oedema, the absence of primary closure of interdental spaces, and bleeding . With inPreservation of intact papilla via MIST SFA promotes wound stabilisation permitting uneventful tissue formation and maturation .Key strengths of this review included the spectrum of databases searched and the large number of included articles, thereby strengthening our statistical analysis. Additionally, all included articles and their reference lists were thoroughly examined by the authors, and cross-matched against the predetermined inclusion and exclusion criteria, thus enhancing the reliability of this systematic review. Furthermore, the use of the rigorous evaluator “AXIS Tool” for the scrutinization of information presented in primary research articles permitted more in-depth assessment of individual aspects of study design, providing a complete and overall qualitative analysis. This subjectivity permitted greater flexibility by incorporating quality of reporting concomitantly with risk of bias when assessing articles and providing a clear advantage over other appraisal tools such as the “Cochrane Risk of Bias Tool” which fails to address poor quality reportage .Despite our best attempts, some review limitations have been identified. For example, one included study within the meta-analysis was a definitive contributor of result heterogeneity . HoweverOur search focussed on an array of observational and primary studies, literature reviews, and comparative and case studies. Despite our investigation having revealed a link between oral FN species and female-specific BC, to date there is still deficits of current data availability.With this in mind, investigations into this field with more focussed assessment objectives would be useful, thus taking the following into consideration: (i) IIF-“Identification-Isolation- Flag-up” of sub-types of FN species presence within the varying sections of breast tissue from a population of women with confirmed BC at different disease developmental stages (lowest grade-highest grade). This would allow for a precise analysis of microbial population counts and examination of the proportional activity levels of each subtype of FN species present. (ii) The exact same procedure would be implemented in a population of women with confirmed diagnoses of varying oral inflammatory conditions and at varying stages: (A) gingivitis (mild-moderate-severe) and (B) periodontitis (mild-moderate-severe).We strongly believe that multiomics analysis  has immeOur motion would aid DHCP's in the promotion of good oral hygiene maintenance   and coulThis systematic review and meta-analysis imply a significant relationship between oral FN species and their role in female-specific BC pathogenesis, highlighting their biomarker potentiality. Our revelations raise awareness of the intricacies of this relationship with respect to adult female populations and could potentially help to reduce/prevent the risk of BC development worldwide, fundamentally transforming the future of both medical and dental prognostics.


P = 0.0045) and posterior circulation  presented the significantly higher risk of recurrence after coil embolization. Meanwhile, intracranial aneurysms of size > 7 mm  had a significantly higher risk of recurrence after coil embolization. Moreover, ruptured aneurysm  and subarachnoid hemorrhage (SAH)  was positively correlated with the risk of recurrence after coil embolization. In conclusion, this meta-analysis identified the characteristics of intracranial aneurysms with MCA, posterior circulation, size > 7 mm, ruptured aneurysm, and SAH as the risk factors of recurrence after coil embolization for intracranial aneurysms.Intracranial aneurysm is a severe cerebral disorder involving complicated risk factors and endovascular coiling is a common therapeutic selection for intracranial aneurysm. The recurrence is a clinical challenge in intracranial aneurysms after coil embolization. With this study, we provided a meta-analysis of the risk factors for the recurrence of intracranial aneurysm after coil embolization. Nine studies were included with a total of 1,270 studies that were retrieved from the database. The sample size of patients with intracranial aneurysms ranged from 241 to 3,530, and a total of 9,532 patients were included in the present meta-analysis. The intracranial aneurysms that occurred in middle cerebral artery (MCA) (OR = 1.09, 95% CI: 1.03–1.16,  The cerebral arterial aneurysm is a severe and prevalent disorder, which is a leading cause of sudden neurological disability secondary to rupture , 2. IntrIn this study, we were interested in the exploration of the correlation of risk factors, such as gender, smoking, posterior circulation, anterior cerebral artery (ACA), interior carotid artery (ICA), middle cerebral artery (MCA), aneurysms of size, aneurysms of the neck, ruptured aneurysm, and SAH, with the recurrence of intracranial aneurysms after coil embolization.The inclusion criteria were as follows: the study type is a retrospective study; the language is limited to English.Exclusion criteria: duplicate publication; research without full text, research without the needed information of this study, incomplete information, or inability to conduct data extraction; animal experiments; reviews and systematic reviews.In this meta-analysis, we searched Pubmed, Embase, and Cochrane Library from the establishment of the database to December 2020. The search terms are mainly: “Intracranial Aneurysm” “Brain Aneurysm” “Anterior Communicating Artery Aneurysm” “Basilar Artery Aneurysm” “Cerebral Aneurysm” and “Coil embolization” “recurrence”. Keywords were combined with Boolean operators to increase search sensitivity and specificity.The literature search, screening, and information extraction were all independently completed by two researchers. When there were doubts or disagreements, the decision was made after discussion or consultation with a third party. The data extraction included the author, year, study area, research type, number of cases, and the OR and 95%CI of age, smoking, posterior circulation, ACA , ICA (interior carotid artery), MCA , Ruptured aneurysm, SHA (subarachnoid hemorrhage), aneurysm size > 7mm and aneurysm neck > 4mm for the prediction of recurrence after coil embolization for intracranial aneurysms. Aneurysm recurrence was defined as inflow into a previously completely occluded aneurysm or growth of an incompletely occluded aneurysm  .Two authors (GENG GUO and JIANZHONG ZHENG) independently conducted literature quality evaluations using the NOS  for the retrospective study. When the opinions are inconsistent, it is decided through discussion or consultation with a third person. The meta-analysis was performed based on the related items of the Preferred Reporting Items for Systematic Reviews and Meta-analysis statement (PRISMA statement).P ≥ 0.1 and I2 ≤ 50%, it indicates that there is homogeneity between studies, and the fixed effects model is used for combined analysis; if P < 0.1, I2 > 50%, it indicates that the study is heterogeneous, and we use sensitivity analysis to find the source of heterogeneity. If the heterogeneity is still large, we use the random-effects model or give up the combination of results and use descriptive analysis. A funnel plot was used to analyze publication bias.Following eligibility verification, data about the risk factors, such as gender, smoking, posterior circulation, ACA, ICA, MCA, aneurysms of size, aneurysms of the neck, ruptured aneurysm, and SAH, were extracted from the manuscript text, patient demographic tables, and on-line tables and figures. STATA 15.1 was used to analyze the data. Odds ratio  was used to analyze the risk factors of no-reflow/slow-flow. Higgins index (I2) is used to evaluate heterogeneity. If the heterogeneity test is In this study, a total of 1,270 studies were retrieved from the database. After eliminating duplicate studies, 638 were obtained. After browsing titles and abstracts, 385 studies were obtained. Finally, nine studies were meta-analyzed through full-text reading .A total of nine retrospective studies were included in this meta-analysis. The sample size of patients with intracranial aneurysms ranged from 241 to 3,530, and a total of 9,532 patients were included in the present meta-analysis. Patients in four studies were from China, patients in two studies were from Korea, and the others were from Europe and America. The NOS score used for quality assessment is all above seven and meets the requirements. The baseline characteristics quality assessment of the included studies is shown in P = 0.793 > 0.1), a meta-analysis was conducted through a fixed-effects model. The pooled results show that there is no significant association between gender  and recurrence after coil embolization for intracranial aneurysms , thus we found that there is no significant association between smoking and recurrence after coil embolization for intracranial aneurysms  and recurrence after coil embolization for intracranial aneurysms. There are 4 studies, including 4,558 patients, that reported the association between gender  and recurrence after coil embolization for intracranial aneurysms. Since there is no significant heterogeneity , a meta-analysis was conducted through a fixed-effects model. The pooled results indicate that intracranial aneurysms that occur in posterior circulation have a significantly higher risk of recurrence after coil embolization , a meta-analysis was conducted through a random-effects model. The polled results indicate that there is no significant association between ACA and recurrence after coil embolization for intracranial aneurysms  from a random-effects model  from a fixed-effects model , a meta-analysis was conducted through a fixed-effects model. The pooled results indicate that intracranial aneurysms of size > 7 mm have a significantly higher risk of recurrence after coil embolization. , a meta-analysis was conducted through a random-effects model. The pooled results indicate that there is no significant association between aneurysms of neck > 4 mm and recurrence after coil embolization for intracranial aneurysms  and SAH  can all significantly increase the risk of recurrence after coil embolization with random effects models .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.


Obesity is a major risk factor for atherosclerotic cardiovascular disease (ASCVD) and is associated with altered platelet function. The mean platelet volume (MPV) is a rapid measure of platelet activation and a prognostic marker in patients with cardiovascular disease. However, no meta-analysis on the association between MPV and obesity has been conducted, and the value of monitoring the MPV in patients with obesity remains unclear.To provide cumulative evidence on whether the mean platelet volume (MPV) is increased in individuals with obesity and to describe associations between the ASCVD-risk factors and the MPV in individuals with obesity.This meta-analysis was prepared following the Meta-analysis Of Observational Studies (MOOSE) guidelines. We searched the PubMed and Embase database from inception until the 31st of March 2021. Studies were included when they reported the mean platelet volume in individuals with obesity and provided a suitable non-obese comparator group. The risk of bias was independently assessed by two reviewers using the Newcastle–Ottawa scale. The primary outcome of the meta-analysis was the MPV, while we considered the atherosclerotic risk profiles as a secondary outcome.p < 0.001) and directly related to changes in triglyceride levels : 1.14, p < 0.001).We identified 178 citations through the PUBMED and 255 citations through EMBASE database search. In all, 13 studies met the inclusion criteria. Firstly, we report an increased mean platelet volume in individuals with obesity compared to non-obese individuals . Moreover, the reported increase in the MPV was inversely associated with the body mass index : 0.18, This meta-analysis and meta-regression showed an increased MPV in nondiabetic individuals living with obesity. Moreover, the MPV was associated with hypertriglyceridemia, an independent predictor of atherosclerotic cardiovascular disease. Overall, the findings suggest that MPV may be a valuable rapid marker for the monitoring and risk-stratification of individuals with obesity who may be at risk of developing cardiovascular disease.The online version contains supplementary material available at 10.1186/s40795-022-00541-8. Obesity affects more than a third of the population living in low-to-middle- and high-income countries. It has also emerged as a significant healthcare challenge affecting children and adults . On a glA previous meta-analysis of Mendelian randomisation studies provided evidence of obesity as an independent predictor of coronary artery disease and showed no association with the risk of strokes . A hyperChu et al., showed the potential value of the MPV as a prognostic marker in patients with CVD [A clinical dilemma arises in patients with obesity, as obesity is associated with impaired platelet responses. The mean platelet volume (MPV) is a convenient measure of platelet activation that is routinely available and can be utilised in an inpatient and outpatient setting . In a prwith CVD . The MPVwith CVD . Notablywith CVD . Althougwith CVD , 17–23 awith CVD  MPV in oNo meta-analysis on the MPV in individuals with obesity has been conducted to date. The value of the MPV as a determinant of thrombotic risk in obese patients may offer a practical and cost-effective surrogate of platelet activation in obese patients who may be at risk of developing ASCVD. This meta-regression and meta-analysis were conducted to estimate the MPV in individuals with obesity by pooling data reported in studies evaluating the MPV in obese populations. The emphasis of the current study is to provide cumulative evidence on whether the MPV is increased in individuals with obesity and to describe associations between the ASCVD-risk factors and the MPV in individuals with obesity.The study protocol was not registered, and therefore, no registration number has been allocated. This systematic review and meta-analysis was reported in accordance with the Meta-analysis Of Observational Studies in Epidemiology (MOOSE) guidelines . All datIs the mean platelet volume increased in individuals with obesity?Are there any associations between the atherosclerotic cardiovascular-risk factors and the MPV?st of March 2021. The search was independently conducted by two reviewers (BBN and VM) using the PUBMED and OVID interface. The reviewers used the following MeSH terms and text words to retrieve the relevant studies; “Obesity” AND “Blood Platelets” OR “Thrombocytes” OR “Platelet count” OR “Mean Platelet Volume” OR “Plateletcrit” OR “Platelet Distribution Width”. The search strategy was adapted for the ISI Web of Science, the international prospective register of systematic reviews (PROSPERO), Cochrane Collaboration, and the Joanna Briggs Institute (JBI) protocol registry. We restricted the search to human clinical studies with no language restrictions applied. Additional scanning of the bibliographies of the retrieved studies was used to augment the retrieval strategy and ascertain that no relevant citations were inadvertently filtered out through the database search. A detailed description of the search strategy used on the MEDLINE database using the PUBMED search engine is provided on supplementary file We conducted a systematic search using medical subject headings (MesH) on MEDLINE and EMBASE search headings (Emtree) on the EMBASE database, from inception until the 31The studies were included based on the fulfilment of the following eligibility criteria:For adult participants, we included nondiabetic individuals living with Class I obesity (BMI 30.0-34.9 kg/m2 ) and class II obesity (BMI 35 35.0-39.9 kg/m2). Whereas for Asian populations, adults with a BMI ≥ 25 kg/m2 were considered as living with obesity.For children and adolescents, we included nondiabetic individuals living with Class 1 obesity (BMI ≥ 95th percentile to <120% of 95th percentile for age and sex) and class II obesity (BMI ≥ 120% to < 140% of 95th percentile or BMI ≥ 35 kg/m2).There were no specific interventions considered for this study. While the comparator included non-obese individuals (BMI < 25.5).The primary outcome included the difference in mean platelet volume. The secondary outcome included atherosclerotic cardiovascular disease (ASCVD) risk markers, such as BMI; systolic blood pressure; diastolic blood pressure; high-density lipoprotein and low-density lipoprotein cholesterol.In this systematic review and meta-analysis, we included observational studies. We included studies that reported the mean platelet volume as one of the study outcomes. Studies that were case studies, reviews, letters to the editor, animal models or cell lines were excluded.Two reviewers (BBN and VM) independently extracted detailed study-level participant data using the adapted Cochrane Consumers and Communication Review Group data extraction sheet for included studies template . The datThe Newcastle–Ottawa tool was used to assess the quality of the included studies . The too2 > 50%), and we used a fixed-effects model when the levels of statistical heterogeneity were low (I2 < 40%) [2 and I2 statistics. Since platelet indices may be affected by geographical variations, a post hoc subgroup analysis was performed based on the reported study location. Moreover, publication bias was assessed using visual inspection of funnel plots and the Egger’s test. All statistical analyses were performed using SATA 16.0 . The risk of bias plots were prepared using the robvis tool [We computed and estimated the standardised mean difference (SMD) and 95% confidence interval (CI) between obese and non-obese individuals for each included study. The reported probability values  were also pooled using Edington’s additive method . We then2 < 40%) . The betvis tool .We performed a random-effects-weighted meta-regression analysis to assess the association between BMI and mean platelet volume. Due to known changes in MPV levels due to age and BMI, a meta-regression was used performed to explore the potential interaction between age and the MPV. The residual maximum likelihood method was used to calculate the between-study component of variance.n = 9), Poland (n = 1) [n = 1) [n = 1) [n = 1) [p < 0.01) with higher SBP  and DBP  when compared to controls. The characteristics of the included studies are presented in Table We identified 178 citations through the PUBMED, 255 through EMBASE database search and two citations through searching registries. We screened all retrieved citations, and after excluding duplicates, a total of 430 studies were eligible for full-text screening. In all, 13 studies were included in the qualitative synthesis, whereas eleven studies were included in the meta-analysis  , Malaysi [n = 1) , China ( [n = 1) , and Brap > 10%), but they also fell in the region of statistical significance (p < 5%). This suggests that asymmetry was a result of alternative sources but not solely a result of publication bias. In addition, we performed the Egger's regression test to explore the association between the reported effect size and the study size. In addition, there was no evidence of small study effects (p = 0.34) (Supplementary table p = 0.51).There were 11 studies included in this meta-analysis, and visual inspection of the funnel plots indicated asymmetry. The contour-enhanced funnel plot was used to distinguish between publication bias and alternative sources of asymmetry. The funnel plot revealed that small studies were not only found in the regions of no statistical significance , which were unexplained by the test for subgroup analysis based on the risk of bias in the included studies ; total cholesterol ; LDL cholesterol  and triglycerides   of the included studies reported on fasting blood glucose levels –20, 22, p = 0.19). This suggests that the reported differences in MPV may be influenced by the risk of bias in the included studies.There were substantial levels of statistical heterogeneity in the reported effect estimates. We, therefore, performed a subgroup analysis to explore the sources of heterogeneity based on the differences in the risk of bias in the included studies Fig. . Althoug2 = 94.1%) when compared to studies with a moderate-quality . In comparison, studies with a low-quality showed the highest increase in MPV in individuals with obesity compared to controls . Notably, amongst the studies reporting on a cohort of individuals with obesity from the Middle East, all studies were conducted in Turkey, and only 28.6%(n = 2) of the studies were moderate-quality. While the majority, 71% (n = 5) of the studies had a low-quality of bias  when compared to studies with a moderate-quality of bias  is congruent with the previous study by Esen et al. [p < 0.001).In the meta-analysis, the levels of MPV were significantly elevated in obesity with moderate levels of certainty Table . Interesn et al. , these cn et al. , 22, 23.This meta-regression and meta-analysis is the first to estimate the MPV in individuals with obesity by pooling data reported in studies with a primary or secondary aim of evaluating the MPV in obese populations. The risk of bias in included studies was low to moderate, with only 15% of the studies deemed high-risk. In addition, the quality of the evidence was moderate for the primary outcome aimed at providing an estimate of the MPV in individuals with obesity. In comparison, the quality of evidence regarding the secondary outcome of ASCVD-risk in individuals with obesity was high. The MPV remains a cost-effective measure of platelet activation, which is under-utilised in the clinical setting. The primary reasons include methodological reasons and a lack of definitive reference intervals associated with clinical significance or outcomes. In our study, the majority of the included studies reported on EDTA collected blood (46%), 32–34, Our primary findings demonstrate the MPV is moderately higher (MD: 0.79) in individuals with obesity. Notably, substantial levels of statistical heterogeneity remained unexplained following a subgroup analysis based on the levels of potential bias and sensitivity analysis based on the age of the included participants. The secondary outcome of this study was addressed by performing a meta-regression, aimed at assessing the associations between the MPV and ASCVD-risk in individuals with obesity. The reported pooled estimates in this meta-analysis are congruent with previous findings of elevated MPV in individuals with obesity , 32, 34.Upon activation, platelets undergo degranulation and change shape from quiescent disks to spread spheres with an increased diameter which correlates to the measured MPV. Larger platelets are metabolically active and densely express fibrinogen receptors, increasing their functional capability and affinity to forming aggregates . IncreasOur secondary findings from the meta-regression analysis aimed at evaluating the associations between the ASCVD-risk factors and the MPV showed a strong association between MPV and the differences in the ASCVD-risk profiles of individuals with obesity compared to controls. An elevated MPV is associated with coronary artery disease in patients with the metabolic syndrome . Notablyp = 0.004) (Table . showed that the MPV was inversely associated with fasting triglyceride levels in non-obese individuals [In our meta-regression analysis, the mean differences in age were associated with a decreased MPV Table . Unlike ividuals .Table 4MThe strengths of the meta-analysis and meta-regression include the comprehensive search and levels of certainty with minimal risk of bias in the reported findings. Although the interpretation of these findings should be taken with caution, the geographic distribution of the included participants was largely restricted to European countries, with mainly studies reporting on individuals with obesity in Turkey. Hence the potential influence of geographic settings on the reported effect estimates could not be determined using a subgroup analysis based on geographic regions due to the few studies from South America and Asia included in this meta-analysis. This would result in misleading subgroup effects due to covariance because of an unequal number of studies in each subgroup. Lastly, all the studies included in the meta-analysis and meta-regression were cross-sectional studies, and these are known to overestimate the overall effect estimate. Future studies should focus on determining cut-off MPV values associated with primary cardiovascular events in individuals with obesity who may have variable responses to antiplatelet therapy. The MPV may be useful in predicting patients at risk of residual platelet reactivity due to increased levels of circulating larger and hyperactive platelets.This meta-analysis and meta-regression provide comprehensive evidence on increased MPV levels associated with increased triglyceride levels and inversely associated with BMI in individuals with obesity. Overall, the findings suggest that MPV may be a valuable rapid marker for the monitoring and risk-stratification of individuals with obesity who may be at risk of developing cardiovascular disease associated with an elevated MPV and hypertriglyceridemia.Additional file 1. Reporting checklist for meta-analysis of observational studies.Additional file 2: Table S1. Search strategy used on the Medline database using the PubMed search engine. Table S2. Search strategy used on the Embase database using OVID interface. Table S3. Assessment of risk of bias using the Newcastle-Ottawa Scale. Table S4. Analysis of publication bias and of small-study effect in the included studies. Table S5. Sensitivity analysis of studies included in meta-analysis of studies reporting on the MPV in obese individuals


Protein- and enzyme-related genes that were found to be associated with PJI included: MBL  and MBL2 . The only receptor-related gene found to be associated with PJI was VDR . This review compiled a variety of genetic polymorphisms that were associated with periprosthetic joint infections. However, the power of these studies is low. More research must be conducted to further understand the genetic risk factors for this serious outcome.The number of orthopedic procedures, especially prosthesis implantation, continues to increase annually, making it imperative to understand the risks of perioperative complications. These risks include a variety of patient-specific factors, including genetic profiles. This review assessed the current literature for associations between patient-specific genetic risk factors and perioperative infection. The PRISMA guidelines were used to conduct a literature review using the PubMed and Cochrane databases. Following title and abstract review and full-text screening, eight articles remained to be reviewed—all of which compared single nucleotide polymorphisms (SNPs) to periprosthetic joint infection (PJI) in total joint arthroplasty (TJA). The following cytokine-related genes were found to have polymorphisms associated with PJI: TNFα ( Orthopedic surgeries, such as ACL repair and primary total arthroplasties, register among the most effective means to treat musculoskeletal disease and injury. The incidence of these orthopedic procedures has grown with projections demonstrating increased demand in the coming years ,2,3. In Periprosthetic joint infection (PJI) is associated with longer hospital stays and increased costs, primarily due to the preferred course of treatment: revision surgery ,7. With A genetic polymorphism is a variation in the genetic code of a particular gene that occurs in more than 1% of the population. Genetic variants include single nucleotide polymorphisms (SNPs), frame-shift mutations, chromosomal abnormalities, translocations, deletions, and duplications. Many genetic polymorphisms associated with PJI following TJA have been identified, particularly those related to antithrombin, protein C, protein S, factor V Leiden mutation, and thrombin ,22,23,24This systematic review and meta-analysis were conducted according to the preferred reporting items for systematic review and meta-analysis (PRISMA) guidelines and the Cochrane handbook.A comprehensive literature search was performed using the PubMed and Cochrane databases. To ensure inclusion of all available evidence, we also manually searched the references of previous studies for other studies which met our inclusion criteria. The search strategy was a combination of subject headings and free text words, and the following MeSH term keywords were used: genetic polymorphism; implantation, joint prosthesis, total hip arthroplasty, complication, joint revision, orthopedic, postoperative, genome wide association study, and genetic susceptibility.Inclusion and exclusion criteria were derived from population, intervention, comparison, and outcome (PICO) and non-PICO based exclusion taxonomy. Inclusion criteria were defined as any publication measuring association or causation between patient genetic risk factors and periprosthetic infection in any orthopedic surgery. Genetic risk factors included single nucleotide polymorphisms (SNPs), frame-shift mutations, chromosomal abnormalities, translocations, deletions, and duplications. Orthopedic surgeries included total shoulder (TSA), total knee (TKA), and total hip arthroplasties (THA). Exclusion criteria included comparisons of any non-genetic risk factors to perioperative infection, or comparisons of genetic risk factors to any non-infection perioperative complication. There were no exclusions based on study location, surgical approach, type of material used for implant fixation, prosthesis type, or other differences in surgical procedures.Two reviewers  independently screened abstracts and full-text articles derived by the search strategy and selected eligible articles based on the inclusion criteria. Reviewers then jointly addressed differences in the outcomes of the independent screenings.Two independent reviewers  extracted information from all included articles. A data collection spreadsheet was created to sort quantitative and qualitative data for analysis. The data were extracted using the following variables: (1) demographics and characteristics , (2) type of surgery, (3) thromboprophylaxis, and (4) genetic polymorphism and gene detection. In addition, primary outcome variable (incidence of PJI events) was also extracted.p-values less than 0.05. Characteristics of the studies were compiled and reported.Two reviewers  independently applied an 11-item quality checklist, derived from the STREGA (Strengthening the Reporting of Genetic Association Studies) and STROBE  checklists to evaluate the methodological quality of the included studies. This 11-item quality checklist includes: (1) clear statement of objectives and hypothesis, (2) clear eligibility criteria for study participants, (3) clear definition of all variables, (4) clear diagnostic criteria, (5) replicability of statistical methods, (6) credible genetic testing method, (7) assessment of Hardy–Weinberg equilibrium, (8) assessment of ethnicity, and addressing the problem of mixed ethnicities statistically (if applicable), (9) sufficient descriptive data , (10) statement of genotype frequencies, and (11) consideration of population stratification. Statistical significance was represented by http://www.genemania.org, accessed on 1 August 2022) provides a flexible, user-friendly analytics web interface for generating hypotheses based on gene functions, analyzing gene lists and prioritizing genes for functional assays. We adopted GeneMANIA to construct a gene–gene interaction network for genetic interaction involved in the incidence of PJI in terms of physical interactions, co-expression, predictions, co-localization, and genetic interaction, as well as to evaluate their functions.GeneMANIA , the fixed effects model was used to estimate the overall effects. If there was significant heterogeneity among included studies, the random effects model was used. The meta-analysis was performed using RevMan 5.3 software.A total of 299 manuscripts from 1997 to 2019 were retrieved via Pubmed (287) and Cochrane (12) databases during the initial identification phase . Of thesOf the eight included studies, seven are case-control studies and one is a prospective cohort study. Publication years ranged from 2006 to 2018, and all studies evaluated the association between PJI and SNP expression, of which 27 SNPs were reviewed. A total of 2308 patients were enrolled in these eight studies. Fifty-nine revision TKAs, 29 revision THAs, 1916 primary THAs, and 367 primary TKAs were included. The overwhelming majority of studies (7/8) were conducted with Caucasian populations, while one study was conducted with a Turkish population. The mean age at surgery for PJI and control patients across all studies was 66 and 58 years of age, respectively. Included studies were published in the following countries: Czech Republic (4/8), United Kingdom (3/8), and Turkey (1/8), as seen in Based on the 11-item quality checklist derived from STREGA and STROBE, the studies included in this meta-analysis contained an average score of 10.25, with a median of 11, mode of 11, and range of 7 to 11. Most of the studies fulfilled all the requirements to generate a strong quality assessment score, including stating the objectives and hypothesis, providing clear eligibility criteria, defining all variables, containing replicable statistical methods, providing sufficient descriptive data , and stating genotype frequencies. The descriptive statistics for each criterion of the STREGA and STROBE methodology score are shown in The gene-gene interaction network generated by GeneMANIA identified seven genes linked to PJI: mannose binding lectin 2 (MBL2), colony stimulating factor 3 (CSF3), interleukin receptor 1 antagonist (IL1RN), vitamin D receptor (VDR), tumor necrosis factor (TNF), interleukin 6 (IL6), and interleukin 1B (IL1B). An additional 20 peripherally linked genes were also identified and can be found in Of the eight studies, minimal overlap existed among the specific genes analyzed across all studies. Therefore, only statistically significant findings for genes analyzed once are reported below.p < 0.0006), the 1/2 and 2/2 alleles in IL-1 RN-VNTR, IL-6, and CT genotype in GCSF3R (p < 0.02) with PJI, but no association between IL-8 and IL-17 with PJI [Four studies reviewed twelve genes encoding for cytokines and cytokine receptors  in association with the development of PJI ,20,22. Ep < 0.01) and C/C genotype (p < 0.05) frequency at position 550 increased risk of PJI. G/G genotype frequency at Codon 54 SNP also increased risk (p < 0.05). No significant associations between codon 52 SNP or promoter 221 SNP and PJI were discovered [p < 0.010) and septic group (p < 0.016) [Malik et al., in a case-control study of 162 revision THAs and 150 controls, found significant associations between MBL SNPs and PJI . In the scovered . NavratiMalik et al., in a case-control study of 162 THA’s  and 150 controls, found no association between MMP1-1, MMP1-3, or MMP1-4 and PJI .p < 0.007) and T/T genotype (p < 0.028) [Mrazek et al., in a case-control study of 350 TJA patients  and 189 non-TJA controls, found no differences in TLR2 or TLR4 allele frequencies between the PJI group and both the aseptic failure and control group . Vitamin< 0.028) .Malik et al., in a case-control study of 162 THAs  and 150 controls, found no significant association of SNPs in OPG-163, OPG-245, OPG+1181, or RANK+575 and PJI when compared to aseptic controls . In agreThe C allele and C/C genotype of IL6-174 were the only polymorphisms applicable to meta-analysis, as they were the only polymorphisms studied across multiple studies.p < 0.035) and PJI [p = 0.67), as shown in Three studies examined the association of the C/C genotype of IL6-174 and PJI following TJA. Erdemli et al. reported significant associations between the C allele at position 174 in IL-6 , as shown in Two studies examined the association of the C allele of IL6-174 and PJI following TJA. Both Malik et al. and Stahelova et al. report no significant association between the C allele polymorphism and the incidence of PJI following TJA ,22. Our Perioperative infection, particularly periprosthetic joint infection, is a major cause of prosthesis failure, morbidity, and mortality in patients undergoing orthopedic procedures. There are several biological and environmental risk factors that predispose patients to perioperative infection including: BMI, ASA scores, diabetes mellitus, alcohol use, and heart disease . In addiVariability in genes that encode cytokine mediators may affect their transcription levels and consequently the innate immune response, therefore altering susceptibility to PJI . This reSimilarly, all the associations found between other significant genes and PJI had small, primarily Caucasian patient populations, which made conclusions less generalizable to the American population. Moreover, there were less than two publications examining any single SNP. In addition, data from these studies were unavailable for further statistical analyses that may have expanded upon the significance of identified associations.This review did have limitations. There was publication bias, as it did not include articles written in languages other than English. Additionally, although a meta-analysis was conducted, this was performed using allele frequencies in septic vs. aseptic groups. Original data from the study authors were not extracted. Seven of the eight studies also assessed primary TJA, while only one assessed revision TJA. Theoretically, the proportion of individuals with these polymorphisms would be higher in the revision TJA group. Therefore, if a patient has septic loosening, screening for these polymorphisms could potentially have higher implications. This should be considered in future studies.Advances in genetics have led to more than 4400 genes being identified as disease-related, with this number steadily increasing . This haThere are several SNPs in genes important to innate immunity that increase the risk of periprosthetic joint infection. TNFα-238 A allele, GCSF3R CT genotype, IL1 RN-VNTR 1/2 and 2/2 alleles, IL1B-511 T allele, and MBL C allele demonstrate increased risk of perioperative joint infection. Future studies should re-examine these SNPs to build a more robust body of evidence on their possible detrimental effects. These studies should have larger, more diverse populations studied in a prospective manner. Additional effort must be taken to share data between researchers for further statistical analyses. Phenomenal strides have been made in the field of genomics, becoming more intricate while more readily available and inexpensive. This provides a synergistic opportunity to create biobanks that utilize artificial intelligence to drive in a new era of individualized, patient-specific care.


To systematically evaluate the impact of diabetes on the prognosis of bladder cancer patients after radical cystectomy (RC).  PubMed, Embase, and Cochrane Library databases were selected from inception to October 2021. The studies on the effects of diabetes on bladder cancer patients after RC were included for analysis. The inclusion and exclusion criteria were independently selected for literature screening, the quality of the included studies was evaluated, and data were extracted. P < 0.001) and the risk of tumor-specific death . Sensitivity analysis shows that the stability of this study is well.  A total of 5 cohort studies were included, with a total of 2 661 subjects, including 391 cases in the diabetic group, non-diabetes. Meta-analysis results show that diabetes increases the overall risk of death in patients after RC (HR = 1.36, 95% CI: 1.30 ∼ 1.43,  Diabetes was an independent risk factor in terms of overall and cancer-specific survival in patients who underwent RC. Bladder cancer is one of the most common malignant tumors of the urinary tract, and its incidence is higher in men than in women . More thDiabetes is the most common chronic metabolic disease in the world, and it may have an important impact on the pathogenesis, progression, and prognosis of bladder cancer through various mechanisms , 7. The The databases including PubMed, Embase, and Cochrane Library were all screened for related studies from their inception to November 2021. The search terms included “diabetes,” “bladder cancer,” “urothelial carcinoma,” “bladder neoplasms,” “urinary bladder neoplasms,” “cystectomy,” “mortality,” “survival,” and “prognosis.”Inclusion criteria were as follows: (1) a cohort study with bladder cancer patients as the research object; (2) pathologically diagnosed patients with high-risk non-muscle-invasive bladder cancer or muscle-invasive bladder cancer, and the selected treatment method is RC; (3) in the diabetes group, the diagnosis of diabetes should be before RC; (4) outcome indicators include overall survival (OS) or cancer-specific mortality and other related prognostic indicators; (5) the hazard ratio (HR) and confidence interval (95%CI) are provided in the literature or the HR and 95% CI can be extracted from the Kaplan–Meier curve in the literature; and (6) if the same study has published results from different periods, the latest research data will be included.Exclusion criteria were as follows: (1) epidemiological investigation, case-control study, and other research types; (2) research using partial cystectomy or transurethral resection of bladder tumor and other methods of treatment; (3) literature types such as review, medical record report, meta-analysis, conference paper, and so on; (4) studies with poor research quality and without full text; and (5) studies without reporting relevant outcome indicators.Two reviewers independently screened the literature and evaluated the quality of the literature according to the inclusion and exclusion criteria and then cross-compared the extracted study data. The Newcastle-Ottawa Scale (NOS) cohort study scoring table was used to evaluate the quality of the literature, and the literature with a score of ≥6 was considered as high-quality literature and could be included in meta studies.The data mainly extracted according to the purpose of the study included the first author's name, publication time, country, cohort age, sample size, age, gender, follow-up time, HR, and 95% CI for survival risk or death risk.I2 test was used to test the heterogeneity of the included studies. If I2 ≥ 50%, P ≤ 0.10, there was heterogeneity among the studies, and a random-effects model was used for analysis; if I2 < 50%, P > 0.10, there was no significant heterogeneity, and a fixed-effects model was used for analysis. Sensitivity analysis was performed on the results of the meta-analysis, and funnel plots were drawn for publication bias analysis. If there was publication bias, the cut-and-fill method was used to verify whether the publication bias affected the stability of the combined effect size.Meta-analysis was performed by RevMan software, and the HR and 95% CI in each study were pooled. The After the screening of the studies, we got 1245 studies from the databases. After the duplications were excluded, 1097 studies were left. After the assessment of title and the abstract of the studies, 365 studies were left. And we performed the full-text screening, we finally 5 studies in the present analysis –12. The P < 0.001). Also, the funnel plot demonstrated that there was no significant publication bias.As shown in Then, we performed the subgroup analysis based on the analysis type of the included studies. We first pooled the results from the univariate analysis results. As shown in We studied cancer-specific survival next. As shown in Bladder cancer is the most common malignant tumor of the urinary system . In 2018Diabetes is a chronic metabolic disease. Due to the absolute or relative lack of insulin, the blood sugar in the body remains at a high level for a long time. If it is not effectively controlled, it will eventually lead to cardiovascular and cerebrovascular diseases, kidney damage, retinopathy, and peripheral neuropathy . DiabeteIn this study, meta-analysis confirmed that diabetes can significantly increase the risk of death in bladder cancer patients after RC, resulting in poor prognosis of patients. The remainder of this paper is arranged as the following two reasons why diabetes increases the risk of death in patients after RC surgery. (1) The incidence of complications in diabetic patients is higher. Studies have demonstrated that diabetes is an independent risk factor for perioperative life-threatening complications in cystectomy patients and is positively associated with serious complications. The study by Goodenough et al. pointed out that glycated hemoglobin ≥6.5% and perioperative hyperglycemia were associated with an increased incidence of major complications after abdominal surgery, emphasizing that elective surgery should be more reasonably controlled for blood sugar. Mossanen et al. included 57 553 patients with RC surgery in 360 hospitals in the United States. The meta-analysis showed that the increase in complications will lead to an increase in the mortality rate of patients. (2) Diabetes can lead to tumor recurrence and progression. Type 2 diabetes is the most common type of diabetes, and insulin resistance and hyperinsulinemia often exist in the body. Insulin can combine with the insulin receptor on the surface of tumor cells to promote the mitosis of tumor cells.There are few studies on the effect of diabetes on the prognosis of patients with RC after surgery, and there is no relevant meta-analysis conducted by scholars at present, so the topic selection of this study is relatively new, and the quality of the included literature is high. However, this study also has some limitations. First, only 5 cohort studies were included, and only 2 studies reported tumor-specific mortality risk. None of the 5 studies analyzed glycemic control after RC. Second, the sample sizes of the two studies were small, so further large-sample prospective studies are still needed for demonstration and analysis. Finally, this study only retrieved published Chinese and English studies and did not include dissertations, conference papers, and unpublished studies. There may be publication bias caused by incomplete literature inclusion. It is proved that the stability and robustness of the results of this meta-analysis are high.The results of this study prove that diabetes can significantly increase the risk of death after radical cystectomy in bladder cancer patients. For the future improved treatments of bladder cancer patients with diabetes, close follow-up should be performed during the perioperative period and postoperative period to monitor and control blood sugar levels to reduce the risk of related complications.Diabetes was an independent risk factor in terms of overall and cancer-specific survival in patients who underwent radical cystectomy.


Many primary research studies in ecology are underpowered, providing very imprecise estimates of effect size. Meta‐analyses partially mitigate this imprecision by combining data from different studies. But meta‐analytic estimates of mean effect size may still remain imprecise, particularly if the meta‐analysis includes a small number of studies. Imprecise, large‐magnitude estimates of mean effect size from small meta‐analyses likely would shrink if additional studies were conducted (regression towards the mean). Here, I propose a way to estimate and correct this regression to the mean, using meta‐meta‐analysis . Hierarchical random effects meta‐meta‐analysis shrinks estimated mean effect sizes from different meta‐analyses towards the grand mean, bringing those estimated means closer on average to their unknown true values. The intuition is that, if a meta‐analysis reports a mean effect size much larger in magnitude than that reported by other meta‐analyses, that large mean effect size likely is an overestimate. This intuition holds even if different meta‐analyses of different topics have different true mean effect sizes. Drawing on a compilation of data from hundreds of ecological meta‐analyses, I find that the typical (median) ecological meta‐analysis overestimates the absolute magnitude of the true mean effect size by ~10%. Some small ecological meta‐analyses overestimate the magnitude of the true mean effect size by >50%. Meta‐meta‐analysis is a promising tool for improving the accuracy of meta‐analytic estimates of mean effect size, particularly estimates based on just a few studies. Meta‐analytic estimates of mean effect size can be imprecise and overestimate effect magnitude, particularly if the meta‐analysis includes few studies. Here, I use meta‐meta‐analysis  to quantify and correct for overestimation of the magnitude of mean effect sizes in ecological meta‐analyses. The typical (median) ecological meta‐analysis overestimates the magnitude of the mean effect size by ~10%, and some meta‐analyses overestimate the magnitude of the mean effect size by >50%. These BLUPs are shrinkage estimates. Hierarchical random effects meta‐meta‐analysis subtracts out within meta‐analysis sources of variation in effect size, thereby shrinking the predicted mean effect size for each meta‐analysis towards the grand mean. Meta‐meta‐analysis can be thought of as a way of adjusting the estimated mean effect size for each meta‐analysis by using information from other meta‐analyses of other topics. A meta‐analytic mean effect size that is much more positive, or much more negative, than those from other ecological meta‐analyses likely is the product of sampling error. One can think of the BLUP of the meta‐analytic mean effect size as a prediction of what the mean effect size would be for a given meta‐analysis, if many additional primary research studies were conducted.It might seem implausible that one could improve the estimated mean effect size for any given meta‐analysis using information from meta‐analyses of other, unrelated topics. However, the statistical rationale for shrinkage estimation does not depend on whether all the data included in the analysis concern the “same” or “related” topics  ecological meta‐analysis should be shrunk by ~10% in absolute magnitude. Some mean effect sizes—primarily large‐magnitude means from small meta‐analyses—should be shrunk by >50%. Simulated data and an empirical validation exercise give confidence that shrinkage estimates of mean effect size from meta‐meta‐analysis improve on unshrunken means.Using a compilation of effect sizes from 467 ecological meta‐analyses . These BLUPs are shrinkage estimates: accounting for random variation within and among meta‐analyses shrinks the estimated true mean effect size for each meta‐analysis towards the grand mean. Large‐magnitude meta‐analytic means  will tend to experience more shrinkage.z‐transformed correlation coefficient as the effect size measure. All effect size measures are unitless ratios, and so in principle meta‐analyses using different effect size measures could be included in the same meta‐meta‐analysis. I did not do this, for two reasons. First, different effect size measures are unitless ratios of variables measured in different units, and so have different statistical properties and scientific interpretations. Second, it would have been computationally infeasible to include all 467 meta‐analyses in the same meta‐meta‐analysis, at least with the computational resources available to me. Indeed, it proved computationally infeasible for me to conduct a meta‐meta‐analysis just of meta‐analyses using Hedges' d or g as the effect size measure.I conducted two meta‐meta‐analyses: one of all the meta‐analyses in the compilation that used the log‐transformed response ratio as the effect size measure, and one of all the meta‐analyses in the compilation that used Fisher's To quantify and summarize the amount of shrinkage produced by each meta‐meta‐analysis, I conducted separate hierarchical random effects meta‐analyses, each of which estimated variation in effect size attributable to variation among studies, among effect sizes within studies, and sampling error. Each of these separate meta‐analyses provided an unshrunken mean effect size, which can be compared with the corresponding shrunken mean (BLUP) from the appropriate meta‐meta‐analysis.2.3z‐transformed correlation coefficients, each based on a sample of 20 observations. The true mean effect size for each meta‐analysis was sampled randomly from a normal distribution with a mean of 0.1 and a standard deviation of 0.1. The true mean effect size for each study within a given meta‐analysis equaled the meta‐analytic mean, plus a random deviation sampled from a normal distribution with a mean of 0 and a standard deviation of 0.3. Each effect size equaled the mean of the study from which it came, plus a random deviation sampled from a normal distribution with a mean of 0 and a standard deviation of 0.3. The sampling variance for each effect size equaled 1/(20–3). I fit 50 separate hierarchical random effects meta‐analyses, each of which estimated the mean effect size, as well as partitioning the variance in effect size into different sources . I also fit a hierarchical random effects meta‐meta‐analysis that estimated a grand mean effect size, and partitioned variance in effect size into variance among meta‐analyses, variance among studies within meta‐analyses, variance among effect sizes within meta‐analyses, and sampling variance. The BLUPs from this meta‐meta‐analysis are shrunken estimates of mean effect size for each meta‐analysis, that subtract out other sources of variation in effect size besides variation in the true mean effect size among meta‐analyses.I used two approaches to validate whether BLUPs from meta‐meta‐analysis improve on unshrunken mean effect sizes. The first approach used simulated data. I generated simulated data from 50 meta‐analyses, comprised of 5, 10, 20, 30, or 50 primary research studies . Each primary research study reported two effect sizes. The effect sizes were Fisher's The simulation results show that meta‐meta‐analysis can be helpful, but on their own, they do not establish that meta‐meta‐analysis is always helpful. One could change the simulation parameters to produce scenarios in which shrinkage improved the average estimation error very little, and scenarios in which it improved the average estimation error even more. To establish whether meta‐meta‐analysis is likely to be helpful when applied to ecological meta‐analyses, I conducted an empirical validation exercise.The goal of the validation exercise is to check whether BLUPs of mean effect size for small ecological meta‐analyses accurately predict mean effect sizes after many additional studies have been conducted. The validation exercise used 12 of the largest meta‐analyses in my compilation, ranging in size from 114–243 primary research studies. These were all the meta‐analyses in my compilation that used the log‐transformed response ratio as the effect size measure, and that comprised >100 but <300 primary research studies. Using meta‐analyses with >300 primary research studies proved computationally infeasible on the computer available to me. I first conducted a hierarchical random effects meta‐meta‐analysis of all 12 meta‐analyses, using the same model structure and statistical assumptions as for the meta‐meta‐analyses described in the previous subsection. This meta‐meta‐analysis provided BLUPs of the “true” mean effect sizes. Note that these BLUPs are very close to the unshrunk meta‐analytic means, as expected because these 12 meta‐analyses all are very large. Then, I repeated the meta‐meta‐analysis 12 times, each time omitting all but the first 10 published studies from one of the 12 meta‐analyses. This procedure converted one of the 12 large meta‐analyses into an artificial small meta‐analysis. The BLUPs for these small meta‐analyses are shrinkage estimates of the true meta‐analytic mean effect sizes, informed by data from the other, larger meta‐analyses included in the meta‐meta‐analysis. Finally, I also conducted 12 separate meta‐analyses, each using just the effect sizes from the first 10 published studies comprising one of the 12 large meta‐analyses. These 12 separate meta‐analyses provided unshrunken estimates of the true mean effect sizes. I compared the shrunken and unshrunken estimates of the true mean effect sizes to the mean effect sizes from the meta‐meta‐analysis of all 12 large meta‐analyses.I fit all meta‐analyses and meta‐meta‐analyses reported in this paper via restricted maximum likelihood (REML), assuming a compound symmetric variance–covariance matrix. Meta‐meta‐analyses additionally assumed a sparse variance–covariance matrix. I conducted all statistical analyses using R 3.6.3 running within R Studio 1.3.1093 . The distribution of shrinkage is skewed; the majority of meta‐analytic means shrink very little, but a minority shrinks substantially  = 0.31 up to exp(2.52) = 12.43. BLUPs for meta‐analyses using Fisher's However, although heterogeneity among meta‐analyses does not invalidate their inclusion in the same meta‐meta‐analysis, it does limit the amount of shrinkage that the meta‐meta‐analysis will produce. Meta‐meta‐analysis shrunk the mean effect size for the typical (median) ecological meta‐analysis by just 10%—a fairly modest amount of shrinkage for most purposes. One reason for that modest shrinkage is the substantial heterogeneity in mean effect size among meta‐analyses. Because of this substantial heterogeneity among meta‐analyses in their true mean effect sizes, one cannot greatly improve the average estimation error by shrinking all of the means towards the grand mean . The optimal amount of shrinkage therefore will be fairly small . Meta‐meta‐analysis will not shrink small means very much. Second, hierarchical random effects meta‐meta‐analysis assumes that future studies on any given topic will sample effect sizes from the same (possibly heterogeneous) distribution as previous studies. This assumption usually holds but not always. Approximately 3%–5% of ecological meta‐analyses exhibit true directional trends in mean effect size over time as more and more studies are published ; data curation (lead); formal analysis (lead); funding acquisition (lead); investigation (lead); methodology (lead); project administration (lead); resources (lead); software (lead); supervision (lead); validation (lead); visualization (lead); writing – original draft (lead); writing – review and editing (lead).


At present, the relationship between serum homocysteine (Hcy), fibrinogen (FIB), lipoprotein-a (LPa), and PAD is uncertain, and there has been no meta-analysis to establish the dose–response relationship between their exposure levels and PAD.Relevant literature published in PubMed, Embase, and Web of Science was retrieved. The robust error meta-regression method was used to assess the linear and non-linear dose–response relationship between exposure level and PAD risk. A total of 68 articles, involving 565,209 participants, were included. Combined with continuous variables, the serum Hcy, FIB, and LPa levels of PAD patients were significantly higher than those of healthy individuals. The odds ratios (ORs) of PAD for individuals with high Hcy, FIB, and LPa levels compared with those with low levels were 1.47, 1.14, and 1.76, respectively. The study also showed that circulating Hcy, FIB, and LPa were significantly elevated in patients with PAD compared with controls. The level of Hcy and the risk of PAD presented a U-shaped distribution. The nonlinear dose–response model showed that each 1 μmol/L increase in serum Hcy increased the risk of PAD by 7%. Similarly, for each 10 mg/dL FIB and 10 mg/dL LPa increases, the risk of PAD increased by 3% and 6%, respectively.This meta-analysis provided evidence that elevated Hcy, PIB, and LPa levels may increase the risk of PAD, and the risk of PAD increases with the increase in serum exposure within a certain range. By controlling Hcy level, the incidence of PAD may be reduced to control the PAD growing epidemic.https://www.crd.york.ac.uk/prospero/Trial registration number: PROSPERO (CRD42021250501), The online version contains supplementary material available at 10.1186/s40001-022-00870-1. Peripheral arterial disease (PAD) is characterized by atherosclerosis of the lower limbs. It is the third most common manifestation of atherosclerotic vascular disease after coronary artery disease (CAD) and stroke. PAD affects individuals all over the world , 2. At pAlthough the etiology and pathogenesis remain unclear, PAD is considered a multifactorial disease; the pathophysiological factors causing peripheral arterial occlusion are complex, and atherosclerosis is the main pathophysiological basis of PAD . The forReducing Hcy level may decrease the risk of PAD, and Hcy level has prognostic significance. However, the association between a mildly elevated Hcy level and risk of PAD is controversial. Although the relationship between circulating FIB and PAD has widely been reported in previous original studies and meta-analyses, inconsistent results have also been obtained in recent studies. For example, Small et al. determined hemostatic factors and their contribution to PAD; they suggested that circulating FIB level was not associated with PAD . SeveralEvaluating whether there is a dose–response relationship between certain exposure levels and disease outcomes is an important task of epidemiological research. It provides strong evidence for pathogenic inference and can promote the exploration of disease prevention and treatment measures. Therefore, this dose–response meta-analysis evaluated the relationship between serum FIB, Hcy, LPa and PAD, which may be of great significance for identifying the risk of PAD and improving the clinical course of PAD patients. We conducted this updated overall and dose–response meta-analysis to further explore the association between circulating biomarkers and risk of PAD.www.crd.york.ac.uk/PROSPERO) as CRD CRD42021250501 on June 1, 2021. We carried out the meta-analysis in accordance with the Preferred Reporting Items for Systematic Reviews and Meta-analyses (PRISMA) guidelines [This systematic review was registered at PROSPERO  framework. Taking Hcy as an example, PICOS were included based on the following selection criteria:Population: Patients had to have an ankle–brachial index (ABI) value of ≤ 0.90 with intermittent claudication or asymptomatic PAD or chronic limb ischemia, and without diabetes, chronic renal failure, or metabolic syndrome.Intervention/Exposure: The level of fibrinogen had to be measured according to the Clauss method; the level of total homocysteine (free and protein bound) had to be determined by fluorescence polarization immunoassay; and Lp(a) concentration had to be determined by ELISA.Control: Healthy people had to be without PAD. Control individuals had be from the same geographic region, and the same exposure measurement methods had to exist between cases and controls.Outcomes: OR, mean ± SD.Study design: Cohort study, case–control study, and cross-sectional study.The eligibility criteria were in accordance with the Title/abstract screening: The study had to investigate the association between Hcy and PAD.Full-text review: (1) the study had to be designed as an epidemiological study ; (2) the research had to report the Hcy levels of cases and controls and the standard deviation that we were able to estimate, or to report odds ratios (ORs), relative risks (RRs), or hazard ratios (HRs) with 95% confidence intervals (95% CIs) for Hcy and PAD; (3) patients had to have asymptomatic PAD, intermittent claudication, or chronic limb ischemia, and they did not have diabetes, chronic renal failure, or metabolic syndrome; (4) control individuals had to be healthy individuals without PAD from the same geographic region and matched with patients with respect to age, gender, and presence of type 2 diabetes mellitus.Study selection was performed by two independent reviewers (H. W. and P. W.), and any discrepancies were resolved through discussion with a third reviewer (Y. H.).We extracted the following information from each study: author; country; year; ethnicity; study design; number of control individuals and PAD patients; gender ratio of controls and PAD patients; exposure level; ORs and 95% CIs for the highest vs the lowest level of the exposure variables; ORs and 95% CIs for different levels of the exposure variables; adjusting for confounders. As randomized control trials were not retrieved and only four articles from all the databases involved cross-sectional research, the Newcastle–Ottawa Scale (NOS) guidelines were used to evaluate the quality of the literature in this study. The NOS guidelines-modified studies that achieved six or more stars were considered to be of high quality, otherwise they were marked as low-quality studies.The studies including the serum levels of Hcy, FIB, and LPa of the PAD group and the control group were used to analyze the differences in serum Hcy, FIB, and LPa between patients with PAD and healthy individuals. We combined continuous variables into standard mean difference (SMD) and weighted mean difference (WMD). To find the relationship between the levels of exposures  and the risk of PAD, the summarized ORs and 95% CIs were assessed by random-effects models. Because of the low incidence of PAD  , risk raA potential linear and nonlinear dose–response relationship of Hcy, FIB, and LPa with the risk of PAD was examined by robust error meta-regression (REMR) approach described by Xu and co-workers, namely, we used inverse-variance weighted least squares (WLS) regression with cluster robust error variances  and we cQ test and I2; P < 0.1 and I2 > 50% indicated high heterogeneity between the studies. If there was a conflict between the Q statistic and the I2 statistic, the I2 statistic prevailed. Potential publication bias was assessed by Egger’s test and Begg’s test, with P < 0.05 indicating publication bias. To judge the robustness of the meta-analysis results, we carried out sensitivity analyses, including changing the effect model, trim-and-fill method, and analyses with excluding one study at a time.Statistical heterogeneity was assessed using the The process of literature screening is presented in Fig. The detailed characteristics of the included studies are summarized in Table I2 = 81.6%, P < 0.001), i.e., the serum Hcy concentration of PAD patients was significantly higher than that of the controls  and Egger’s (P = 0.207). Three studies were added with trim-and-fill method. The results did not change significantly, suggesting that the combined effect size results were robust. We also examined the impact of a single study on the results; the removal of any single study did not make a large change in the SMD, i.e., the result of the combined effect size was robust , indicating that the serum FIB concentration of PAD patients was significantly higher than that of the controls  and Egger’s test (P = 0.279). Four studies were added with trim-and-fill method; the SMD did not change significantly, suggesting that the combined effect size results were robust. Examination of the impact of a single study on the result revealed that the removal of any single study did not make a large change in the SMD , meaning that the serum LPa concentration of PAD patients was significantly higher than that of the controls  and Egger’s (P = 0.503). Two studies were added with trim-and-fill method and the SMD did not change significantly, suggesting that the combined effect size results were robust. Examining the impact of a single study on the result revealed that the removal of any single study did not make a large change in the SMD . We examined the impact of a single study on the results; removing two studies [Data extracted from 13 articles  that compared the relative risk of PAD between individuals at the top level of Hcy and those at the bottom level of Hcy yielded a summary OR of 1.470 . We examined the impact of a single study on the results; removal of two studies [Twelve studies, including 303,710 participants, were used to evaluate the risk of PAD in individuals with the highest FIB levels compared with those with the lowest FIB levels. The summary OR was 1.142 . The result of the combined effect size was robust, because there was no significant change in the result after applying the trim-and-fill method, and the removal of any single study did not significantly change the result of the combined effect size . The level of Hcy and the risk of PAD presented a U-shaped distribution  were included to analyze the dose–response relationship between Hcy and PAD. The data were fitted using an RCS with three knots  allowing for a potential nonlinear relationship. The regression parameter estimates of the first spline and the second spline were − 0.4353 (β1) and 0.4180 (β2), respectively (Table P = 0.882). Nonlinear dose–response analysis showed that the risk of PAD continued to increase as FIB increased  were used to analyze the dose–response relationship between FIB and PAD. An RCS was created , which generated two splines, and these were then employed for the potential nonlinear dose-specific modeling. The regression parameter estimates of the first spline and the second spline were 0.0005 (β1) and 0.0020 (β2), respectively  were included to analyze the dose–response relationship between LPa and PAD. There were three knots  using RCSs. Estimated regression parameters were 0.007384 for β1 and − 0.002629 for β2  in patients with PAD compared with controls. Previously, a meta-analysis of 14 relevant studies showed that Hcy was significantly elevated  in patients with PAD compared with controls [U-shaped curve distribution. When the Hcy concentration was lower than 11.7 μmol/L, Hcy was not considered a risk factor for PAD; in contrast, when the Hcy concentration was higher than 11.7 μmol/L, the risk of PAD increased sharply with the increase of Hcy level. This conclusion is supported by the previously reported relationship between PAD and Hcy. [In the present study, we found that the serum homocysteine levels of PAD patients were significantly higher than those of healthy individuals. In addition, the risk of PAD in individuals with high serum homocysteine was 1.47 times higher than that of the corresponding low-level population. Meanwhile, Hcy was significantly higher  levels of PAD patients were significantly higher than those of healthy individuals. In addition, the risk of PAD in individuals with high serum LPa levels was 1.76 times higher than that of the corresponding low-level population. The dose–response relationship between serum LPa level and PAD indicated that as the LPa level increased, the risk of PAD increased. In addition, every 10 mg/dL increase in serum LPa concentration increased the risk of PAD by 6%.The pathogenesis of PAD is complex, involving a variety of genetic and environmental factors related to atherosclerosis and thrombosis, and the interaction between them . The mosOur meta-analysis has several strengths. To our knowledge, this is the first systematic dose–response meta-analysis of serum exposure levels of Hcy, FIB, and LPa and the risk of PAD, and the results obtained have a mutual collaborative relationship with previous research conclusions. Compared with the sample size limit of a single study, this meta-analysis included a total of 68 articles, with a total sample size of 565,209 patients, including > 56,754 PAD cases. The total sample size of the studies on the relationship between Hcy, FIB, LPa and PAD reached 65,886, 320,756, and 255,456, respectively. A large number of cases allowed us to determine the relationship between exposure levels and PAD risk. In addition, the 68 included articles were of high quality . The method used in this study was the REMR method. Compared with previous studies using Generalized Least Squares (GLST) as a dose–response meta-analysis method, the REMR method eliminated the bias generated by the GLST method using the non-intercept model, resulting in better error estimation and a better visual fit to the data . MoreoveSeveral potential limitations should be mentioned in this meta-analysis. First, according to previous reports, the relationship between exposure levels and the risk of PAD may vary slightly by gender and region –38, but Based on the above results, it can be reasonably concluded that serum Hcy, FIB, and LPa are related to the risk of PAD, and within a certain range of their serum levels, the risk of PAD increases with the increase in the serum level. These three exposure factors are expected to become serum biomarkers of PAD. Individuals with high serum Hcy, FIB, and LPa levels should be highly concerned about the risk of PAD, and early screening and appropriate treatment are crucial. By controlling the Hcy level, the incidence of PAD may be reduced to control the growing epidemic. Therefore, our research is of great significance for the prevention of PAD and for improving the diagnosis of the disease. Certainly, further observational studies with large sample sizes are needed to verify our results, and more studies are also needed to explore the underlying physiological and pathological mechanisms.Additional file 1: Table S1. Search strategy.Additional file 2: Table S2. Characteristics of individual studies on Hcy, FIB, LPa and risk of PADs.Additional file 3: Figure S1. Weighted mean difference (WMDs) and 95% confidence intervals (CIs) of the selected studies and the pooled Hcy levels (A), FIB level (B) and LPa level C in patients with PAD and control subjects.Additional file 4: Table S3. Table of relative risks  from the nonlinear dose–response analysis of Hcy, FIB, LPa and risk of PAD.Additional file 5: Table S4. Table of relative risks  from the linear dose–response analysis of Hcy, FIB, LPa and risk of PAD.


The association between intracranial artery calcification (IAC) and the risk of ischemic stroke occurrence or poor prognosis had not yet been fully understood. In this study, we conducted a meta-analysis of existing studies aimed to assess whether IAC can be used to predict future ischemic stroke and post-stroke mortality.Medline, Cochrane, Web of Science and Google Scholar databases were searched up to June 30, 2022. Studies were included if they reported risk ratio (RR) or odds ratios (OR) and corresponding 95% confidence intervals (CI) of stroke concerning the presence of IAC. Random or fixed effects model meta-analyses were performed. Meta-analysis was conducted by using Stata version 16.0.P = 0.001) and stroke recurrence . However, we did not find a significant correlation between IAC and post-stroke mortality .Twelve studies involving 9346 participants were included. Compared with those without IAC, patients with IAC had a higher risk of stroke occurrence (adjusted OR 1.62, 95% CI 1.18–2.23, Our meta-analysis demonstrated that the presence of IAC was identified as an independent risk factor for ischemic stroke occurrence and recurrence but is not a predictor of post-stroke mortality. Stroke has remained in the top three causes of death and a major cause of disability globally, and its absolute number of cases increased substantially from 1990 to 2019 [IAC is known to be an essential risk factor for cerebral infarction and an essential marker of ICAS –5. Sever(CRD42021281357).This study was preformed according to the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) statement. It was registered in PROSPERO international prospective register of systematic reviews We searched Medline, Cochrane, Web of Science, Google Scholar databases from inception to June 30, 2022. Prospective or retrospective cohort and case-control studies were included. Relevant keywords, phrases and medical subject headings (MeSH) terms were used. The following search keywords in different combination were used: stroke, intracranial artery, calcification or calcified or calcium. Furthermore, we hand-searched reference lists of included studies to find additional studies. Only English language studies and accompanied by full-length peer-reviewed papers were included. Two reviewers independently screened and performed parallel assessments of the manuscripts. The following search strategy was used for Web of Science and modified to suit other databases:#1 stroke.#2 intracranial artery.#3 calcification OR calcified OR calcium.#4 #1 AND #2 AND #3.We also used the “cited by” function of Google Scholar to minimize the risk of missing data.Studies screened and examined by titles and abstracts after removing over-lapping and duplicate articles. Only studies that met the following criteria were included in this meta-analysis: limited to human subjects; report risk ratio (RR) or odds ratios (OR) and their corresponding 95% confidence intervals (CI) of stroke relating to IAC, and at least one of the following outcomes should be reported: occurrence, recurrence or mortality. Letters, comments, editorials, case reports, proceedings, and personal communications were excluded.Data extraction was independently performed by two reviewers using a standard form. The data extracted included the following information: first author, study design, publication year, country of origin, sample population demographic characteristics, gender, sample size, endpoint, and covariates adjustment in each study.The quality assessment for observational studies included in this meta-analysis were assessed using the Newcastle-Ottawa Scale (NOS) . This sc2 < 50.0%) or, in the presence of heterogeneity (I2 > 50.0%), random-effects models. Heterogeneity across studies was assessed by using both the Q test and the I2 statistic (ranging from 0% for perfect homogeneity to 100% for extreme heterogeneity). An I2 value > 50% indicates significant heterogeneity. We also performed a subgroup analysis to further assess the association between IAC and the risk of stroke events according to clinical characteristics, sample size and study design. Pooled effects were calculated, and a two-sided P value < 0.05 was generally considered to indicate statistical significance. Publication bias was not performed as enough studies are necessary for this type of analysis. The data analysis was done using Stata 16.0 software.Meta-analysis was performed using either fixed-effects models  –18. The The main characteristics of the included studies were presented in Table p = 0.003; Fig. 2 = 82.3%, P <  0.001). Considering the relatively high heterogeneity, a sensitivity analysis following the leave-one-out approach was performed to show that Kao et al. study had the most significant influence on the heterogeneity, and the pooled RR without this study was 1.75  with the I2 value reduced to 0%. No associations between the modifiers age, study design, sample size, country, study quality and IAC were identified in meta-regression analyses .Compared with those without IAC, among subjects with IAC had an elevated risk of ischemic stroke .Results from three studies demonstrated that the presence of IAC related to a higher risk of stroke recurrence  without significant heterogeneity among the studies  and case-control studies . When stratified by continent, IAC was positively related with increased risk among studies performed in European , North American  and Asian . In the further analysis by the number of cases, studies that included a small sample, that is, < 500 patients  had a higher risk of ischemic stroke than those studies with more than 500 cases . The results of subgroup analyses are shown in table In the study design subtype, a statistically significant effect of IAC on first stroke risk was observed in both cohort studies . As only eight studies assessing the risk of stroke occurrence were included in the present meta-analysis, we could not entirely rule out the presence of publication bias.Visual inspection and Egger’s test indicated an asymmetric funnel plot  and between IAC and stroke recurrence was also found . However, there was no significant correlation between IAC and post-stroke mortality.IAC is a common incidental finding on brain CT scan in the general population and, although most of calcifications are viewed as innocent without any prognostic significance, others are related to adverse clinical outcomes –21. SeveAlthough this study showed that IAC was closely related to the recurrence of ischemic stroke, it must be note that there are still few studies in this fields and the conclusions are inconsistent. In the term of IAC impact on recurrence after ischemic stroke, Wu et al. showed a higher degree of IAC was associated with a high risk of stroke recurrence in patients with cerebral small-vessel disease, which might indicate chronic calcification observed in large intracranial arteries may have potential impacts on the cerebral vascular bed extending to small blood vessels . MoreoveIn our meta-analysis, no significant correlation was found between IAC and post-stroke mortality, still, the investigation of the link between IAC and stroke prognosis may have important clinical significance , 34. AltTo our best of our knowledge, this is the first systematic review and meta-analysis, focus on the relationships between IAC and ischemic stroke. As a common and easily identifiable finding on brain CT, IAC may be a useful indicator to predict the risk and outcome of ischemic stroke. Several limitations of this study should be considered. First, the studies included in this meta-analysis were either cohort studies or case-control studies, thus, the causality between IAC and the risk of ischemic stroke remains unclear. Second, the number of articles included seems too small, therefore, a larger-sized, prospective study is warranted to further investigate the association between arterial calcification and stroke prognosis. Third, we excluded articles that were not published in English to ensure the quality of studies, but it might lead to publication bias. Fourth, it would be very beneficial if subgroup analysis could be conducted based on the arteries involved. Although some recent studies aimed to explore the relationship between different patterns of intracranial arterial calcification and atherosclerotic disease , the exaIn conclusion, the findings from this study indicate an association between IAC and increased stroke risk. Further carefully designed and well-conducted studies with prospective design are needed to conduct to identify our results.