Background: Several researchers have investigated alternative markers related to ovarian responsiveness in order to better predict IVF outcomes, particularly in advanced reproductive-aged women. The follicular output rate (FORT), the follicle-oocyte index (FOI) and the ovarian sensitivity index (OSI) are among the most promising. However, these three metrics have not been investigated as independent predictors of live birth in women of advanced reproductive age; neither have they been compared to the two ‘component’ characteristics that are used to calculate them. (2) Methods: A logistic regression model containing all relevant predictors of ovarian reserve or response was used to evaluate the potential of FORT, FOI and OSI as predictors of live birth. After, the non-linear associations between FORT, FOI and OSI and the probability of live birth were evaluated. Finally, we fitted multiple logistic regression models to compare whether FORT, FOI and OSI were more informative predictors than their components. (3) Results: 590 couples received a total of 740 IVF cycles, after which, 127 (17.5%) obtained a live birth. None of FORT, FOI and OSI showed a strength of association or a p-value even close to female age  1.00 (0.99–1.01), 1.00 (0.99–1.01), 0.98 (0.88–1.11) and 0.58 (0.48–0.72), respectively). The three models comparing FORT, FOI and OSI with the number of oocytes retrieved, the AFC, the number of preovulatory follicles and the FSH total dose were not more informative. (4) Conclusions: In a population of women of advanced age with unexplained infertility, none of FORT, FOI and OSI were predictive of live birth or more predictive than the two ‘component’ characteristics that were used to calculate them. We suggest clinicians and researchers still use female age as the most reliable predictor of an IVF treatment.(1)  The number of retrieved oocytes is considered one of the most important predictors of live birth in an in vitro fertilization (IVF) cycle ,2,3. OocThe individual variability of these intrinsic factors may be associated with an unexpectedly low ovarian response during COS. Thus, the total number of retrieved oocytes does not always accurately reflect the ovarian potential. The wish to find the true ovarian potential in terms of the maximum number of retrieved oocytes and to link this to aforementioned factors has stimulated the research of more sophisticated markers of ovarian function and response, such as the ovarian sensitivity index (OSI) ,8, the fOSI represents ovarian sensitivity to exogenous gonadotropins, and since its introduction, it has been used to adjust the COS regimen in subsequent IVF cycles, whereas FORT and FOI have been claimed to be a good representations of the dynamic nature of follicular growth and follicular competence. As a matter of fact, several studies have shown that these markers correlate positively with IVF outcome ,12,13.However, the aforementioned markers are also affected by some limitations: it is unlikely that all three metrics have simple linear associations with the probability of live birth. OSI does not take into account the gonadotropin regimen, nor does it consider the type of gonadotropin -like activity, or not). Furthermore, it might be misleading if inappropriate low starting doses of exogenous gonadotrophins are given. Neither FORT nor FOI assess the total number of mature oocytes retrieved, whereas both indexes depend on the baseline antral follicle count (AFC). AFC by itself shows limited predictive value in older women, whose follicles are affected by increased granulosa cell apoptosis, impaired mitochondrial function and increased oxidative stress ,15. IndeHowever, no studies have so far considered these three metrics  as independent predictors and compared metrics to the two ‘component’ characteristics that were used to calculate them. In this retrospective study among women aged 39 or above, we first evaluated non-linear associations between metrics and live birth in women of advanced reproductive age, then compared all three metrics to one another, and finally, compared these three metrics to the two variables that were used to calculate them.2) > 32 kg/m2; anti-Mullerian hormone (AMH) < 0.1 and/or early follicular phase follicle-stimulating hormone (FSH) > 20 UI/l; any known cause of female infertility ; and cycles with pre-implantation genetic diagnosis. The BMI criterion is imposed by regional legislation in order to reduce the risk of comorbidities related to obesity in pregnancy. All women included in the study had ovulatory cycles and patency of at least one fallopian tube at sonosalpingography (SSG), and all male partners had normal basic semen parameters according to the indications of the World Health Organization (WHO), 2010.The study was performed in accordance with the Helsinki Declaration and with the approval of the Institutional Review Board. In this retrospective analysis, we included a cohort of women aged 39 years or above and affected by unexplained infertility, who were admitted to the IVF unit of the S. Anna academic hospital between 2010 and 2019. Only autologous cycles were considered. Exclusion criteria were: female body mass index (BMI: Kg/mControlled ovarian stimulation (COS) was performed either with recombinant FSH (rFSH), human menopausal gonadotropin (HP-hMG) or rFSH plus recombinant luteinizing hormone (rLH), under pituitary suppression applying both long and short protocols. The choice of the starting gonadotropin dose was based on age, AMH concentrations and AFC, as well as on response to previous COS. The long protocol was performed by administering buserelin  starting from the late luteal phase of the previous cycle. Pituitary suppression was verified after approximately two weeks  before starting COS. In the short protocol, either the GnRH-antagonist cetrorelix  or ganirelix  was started at a subcutaneous dose of 0.25 mg/d according to a flexible schedule, when at least one follicle ≥12 mm in mean diameter was observed on ultrasound (US).COS was monitored by serial transvaginal US and serum estradiol (E2) measurements performed every second day from stimulation day 6–7. COS continued until at least two follicles reached 18 mm in mean diameter, when ovulation was triggered by injecting either 10,000 international units (IU) of human chorionic gonadotropin (hCG)  or 250 mcg of rhCG  subcutaneously. US-guided transvaginal oocyte aspiration (OPU) was performed approximately 36–37 h after hCG administration, under local anesthesia . Oocytes were immediately recovered from the follicular fluid and then washed in buffered medium and stored until the fertilization procedure.Semen samples were examined to assess sperm concentration, motility and morphology according to the WHO guidelines . The sam2, 6–7% CO2, balanced with N2.Zygotes were placed in pools in 4-well dishes , and embryos were cultured in pre-equilibrated Cleavage Medium (Cook) overlain with mineral oil using incubators  in a gas phase consisting of 5% OAll cleaved embryos were morphologically evaluated under a conventional stereomicroscope using the integrated morphology cleavage score (IMCS) by Holte .Embryo(s) transfer was performed using a soft catheter under ultrasound guidance. According to the policy of our IVF unit during the time period under study, one or two embryos were transferred on day 2/3 post-fertilization. Only embryos reaching the blastocyst stage were vitrified and thawed in subsequent cycles. The luteal phase was supported by administering 180 mg/day natural progesterone  for 15 days.Pregnancy was assessed by serum hCG assay 15 days after embryo transfer (ET) and then confirmed if at least one gestational sac was visualized on transvaginal US after two further weeks.Live birth was defined as the delivery of a live-born infant (>24 weeks of gestation). The cumulative live birth rate (CLBR) was defined as live deliveries (at least one live birth) per women including both fresh and frozen/thawed embryo transfers obtained from the same cycle.FORT was defined as the number of pre-ovulatory follicles of at least 15 mm in response to ovarian stimulation divided by the pre-treatment AFC  multiplied by 100. FOI was defined as the number of oocytes collected after ovarian stimulation divided by the pre-treatment AFC multiplied by 100. For agonist cycles, the AFC was evaluated as previously described .OSI was defined as the number of oocytes collected after ovarian stimulation divided by the total dose of FSH administered, per 1000.First, we fitted a logistic regression model containing all relevant predictors of ovarian reserve or response to evaluate whether these common indicators were ‘independent’ predictors of live birth: female age, duration of infertility, previous assisted reproduction treatment (ART), primary or secondary subfertility, preovulatory follicles, oocytes collected, AFC and total FSH dose. Since only age was found to be a strong predictor of live birth, we evaluated the non-linear associations between the three metrics separately  and the estimated probability of live birth after IVF to obtain an indication of the strength and functional form of the associations. To this end, we fitted logistic regression models on live birth after IVF using restricted cubic splines with 4 knots for the three mentioned metrics and then used these models to predict the probability of a live birth after IVF for a range of values for each metric. We plotted the results using simultaneous 95% confidence intervals (Cis) that take into account the many comparisons drawn in these plots.Next, we fitted multiple logistic regression models to compare whether FORT, FOI and OSI were more informative predictors than the two ‘components’ used to calculate them. For FORT, we compared a model containing FORT to a model containing preovulatory follicles and AFC. For FOI, we compared a model containing FOI to a model containing oocytes collected and AFC. For OSI, we compared a model containing OSI to a model containing oocytes collected and total FSH dose. In addition, the following four factors were added to all models, as they were considered the most important and could explain part of the associations of the three metrics and/or their components: female age, duration of infertility, previous ART and primary or secondary subfertility.We also assessed whether the associations between FORT, FOI and OSI were different for poor responders to stimulation (according to Bologna criteria  defined We determined the best-fitting and most-informative model in terms of the lowest Akaike Information Criterion (AIC) . For allUsing the ‘10 events per variable’ rule of thumb, we would be able to include approximately 11–13 predictors in our model(s) . Using aData were compiled in Excel and analyzed in R 3.6.0 using the rms, mice, xtable and dplyr packages .Data on 590 couples were available. These 590 couples received a total of 740 IVF cycles, after which 127 (17.5%) obtained a live birth. The baseline characteristics are shown in p < 0.001) was a significant predictor of live birth after IVF .The associations between live birth and FORT, FOI and OSI were not significantly different in poor responders than in normal responders . Over the years, several attempts have been made in order to identify surrogate markers of ovarian reserve, which, in turn, could be markers of IVF outcomes. This would allow us to decide for whom treatment is expected to be (cost-)beneficial. However, so far, studies on the association between markers such as AMH/AFC and implantation, pregnancy, and/or live birth after assisted conception have reported conflicting results ,24. Therefore, several authors have investigated alternative markers that are discussed in this paper, particularly those related to ovarian responsiveness to COS, in order to better predict IVF outcomes. Efficient markers would be of particular interest to subgroups of low prognosis patients defined both by the Poseidon and the Bologna criteria ,25. Alviggi et al. analyzed the predictive role of FOI in assessing ovarian sensitivity in hypo-responder patients . The autGrynberg et al. discussed the potential use of FORT as a quantitative and qualitative marker of ovarian responsiveness to gonadotropins, and the possible implications for the applicability of the Poseidon criteria . They stRecently, the amount of hormone medication needed for each oocyte produced  was investigated in a retrospective cohort study that included more than 1200 women undergoing IVF with FSH/hMG stimulation . ConsistSo far, no studies have been published on whether these surrogates offer significantly higher performance than their constituent parameters, particularly in women of advanced age. To be specific: are FORT and FOI more predictive of baseline AFC, the number of pre-ovulatory follicles and the number of oocytes retrieved? As for OSI, is it more informative than the number of oocytes and the total dose of FSH used?Despite the limitations of a retrospective study, our results seem to scale back the capabilities of FORT, FOI and OSI to answer the above questions. The combination of multiple indicators of ovarian reserve and ovarian response to COS (FOI and FORT) or in terms of FSH administered (OSI) does not seem to be more advantageous than the traditional predictors of IVF outcomes. In addition, most of these predictive factors only become available after at least one IVF cycle is conducted, limiting their usage in clinical decision-making about starting IVF. Thus, there is still a knowledge gap regarding the possibility to predict oocyte quality, especially using only information that is available before starting COS. In a cohort of couples with unexplained infertility and advanced female age, only age was found to be a clear predictor of live birth after IVF, further confirming the intrinsic awareness of every IVF expert: the age of the egg is what really matters. This conclusion is in agreement with what has already been reported in previous studies ,28,29.Indeed, numerous factors are implicated in the final outcome of an IVF treatment: oocyte and embryo quality, endometrial receptivity, women’s general health conditions, etc. It is likely that only female age is able to coherently capture all these factors. The impact of age per se seems truly relevant, considering that its correlation with IVF outcome is stronger than that of all markers considered or even combined, in spite of a rather narrow distribution of ages.Note that FORT, FOI and OSI can be considered markers that indicate different aspects of the response to COS; we merely showed that their association with live birth after IVF is negligible when considering female age and their constituent parameters. This result could have more than one explanation. First of all, one should consider that each of the parameters used for the calculation of the three metrics are operator dependent. The follicle count, which registered a reduction in inter-individual variability with the introduction of three-dimensional (3D) technology , is stilAnother reflection should be dedicated to the value of the ‘live birth’ as a definitive goal of an IVF treatment. This argument has been questioned from various points of view . RegardiIn summary, all these parameters suffer from measurement error, offering the possibility of adding more noise, i.e., (non-)random variability. By adding more variables that are operator-dependent in a model, it is not surprising that the accuracy of these surrogates is reduced with respect to the single parameter that underlies it. This could also explain why a simple but perfect measure such as female age seems to be more informative. The main strength of our retrospective analysis is that it represents the first study to investigate whether a non-linear association exists between the metrics FORT, FOI and OSI and LBR in the same homogeneous population, i.e., women of advanced age with unexplained infertility, thus avoiding all the hypothetical confounding factors deriving from all the other plausible causes of infertility . In addition, we carefully assessed the fit of models with the constituent parameters using contemporary statistical methods, adjusting for multiple comparisons and adjusting standard errors for couples receiving multiple cycles.In conclusion, none of the three predictors for fertility that were proposed in the literature  were more predictive than the two ‘components’’ characteristics that were used to calculate them. This was in a population of women of advanced age (39 years or above) with unexplained infertility. Nonetheless, in light of a lack of evidence that backs up using the three metrics, we suggest that clinicians and researchers still use the components themselves  in counseling and prediction modeling.


Endobronchial ultrasound-guided transbronchial needle aspiration (EBUS-TBNA) is less commonly used in nonmalignant diseases. In particular, its application in mediastinal cystic lesions has been reported less frequently. EBUS-TBNA is a reassuringly safe procedure with an overall complication rate less than 2%, and serious adverse event rate of 0.14% to 0.16%. The most common complications are infections .A 28-year-old male presented to the hospital with mediastinal cyst that was incidentally discovered by computed tomography. There was no past history of the patient reviewed.The cyst was identified as a round, anechoic structure by EBUS and serous fluid was aspirated. The carcino-embryonic antigen, mycobacterium tuberculosis DNA and cultures in the fluid were negative. Cytology analysis showed lots of lymphocytes and no malignant cells. The diagnosis of lymphangioma was confirmed based on the computed tomography and EBUS presentation, the nature of the aspirated fluid and the large number of mature lymphocytes within the cystic fluid.Twenty-six hours after EBUS-TBNA, the patient complained of a fever with the highest temperature of 39°C, accompanied by a right-side chest pain, no other symptoms of were reported. The following examinations confirmed the diagnosis of pneumonia, pleurisy, mediastinitis and mediastinal cyst infection, while cultures from cyst and right pleural effusion were both negative. The patient was treated with Teicoplanin+Imipenem/cilastatin, and ultrasound guided transcutaneous catheterization drainage of mediastinal cyst and pleural effusion were performed.Seven days after the treatments, the patient's symptoms resolved, the complete blood count, C-reactive protein, erythrocyte sedimentation rate were lowered. The size of the cyst was slightly reduced on 17 June compared to that before EBUS-TBNA. Although the surgical resection of the cyst was recommended, the patient declined. After extracted the two drainage tubes, the patient was discharged on June 22. The patient was followed up by telephone 6 months after discharge and he remained asymptomatic.EBUS-TBNA is a useful diagnostic and therapeutic tool for the management of mediastinal cysts. However, considering the possibility of serious complications, the clinical procedure should be carried out scrupulously with appropriate patient selection and strict aseptic principles. The application of EBUS-TBNA in mediastinal cystic lesions is usually accompanied with a high incidence of infection.,13,16,25 Herein, followed by a literatue review, we reported a patient with severe infections including pneumonia, pleurisy, mediastinitis and mediastinal cyst infection secondary to EBUS-TBNA, aiming to state the utility and complications of EBUS-TBNA in mediastinal cyst settings.Convex Probe endobronchial ultrasound-guided transbronchial needle (EBUS-TBNA) has been conventionally used for the staging of lung cancer and sampling of mediastinal and hilar lymphadenopathy,2A 28-year-old male patient presented to the hospital with mediastinal cyst that was incidentally discovered by computed tomography (CT) on June 8,2020 Figs. –3. He waTwenty-six hours after EBUS-TBNA, the patient complained of a fever with the highest temperature of 39°C, accompanied by a right-side chest pain, no obvious symptoms of chill, cough, expectoration of sputum and shortness of breath. Repeated chest CT imaging showed that the cyst was smaller than that before EBUS-TBNA, with a size of 74 mm × 45 mm, chest CT also showed pneumonia, pleural effusion of the right side and increased mediastinal density Figs. –11. The 3 About two-thirds of patients with mediastinal cyst remain asymptomatic. However, some patients present with symptoms due to compression of adjacent organs, cyst infection, etc. The common symptoms are chest pain, dyspnea, hemoptysis or cough and expectoration due to recurrent cyst infections. Chest CT or magnetic resonance imaging (MRI) examinations are clinically important for the diagnosis of mediastinal cysts. Typical mediastinal cysts are presented as liquid density (Hounsfield Unit<20) on CT images, mostly round or lobulated, with clear boundaries and no enhancement in contrast CT. However, some infected cysts, proteinaceous, or combined with hemorrhage or calcification have atypical CT manefestations, making them indistinguishable from neoplasms. The accuracy of CT in the diagnosis of mediastinal cysts was reported by Aravena et al to be only 53.8%, about 43% of cysts could be misdiagnosed as solid mediastinal tumors. An MRI is also helpful in the diagnosis of mediastinal cysts, and serous fluid has a low T1-weighed signal and high T2-weighted signal images. For some proteinaceous, hemorrhagic or infected cysts, MRI images are atypical and may show higher T1-weighed signal images. EBUS has unique advantages in diagnosing mediastinal cysts. It may display anechoic (black), hypoechoic (darker), isoechoic (grey), and visible separation in the cyst. The Doppler patterns can be used to differentiate vascular structures. For some cysts with features of solid tumors on CT images due to the presence of proteinaceous content, the fluid-fluid level could be seen on EBUS, which is a more typical feature and can be used to differentiate cysts with features of solid tumors and neoplastic lesions. The fluid-fluid level of noninfected cyst is probably due to the dependent layering of proteinaceous material within the cyst. However, it is difficult to differentiate cysts with tumor necrosis.Mediastinal cystic lesions account for 12% to 18% of primary mediastinal tumors, with bronchial cysts being the most common (40%), followed by pericardial cysts (35%), enterogenous cyst (10%), and unclassified (14%). The origin of the cyst is determined based on its cell components on the cytology specimen: bronchogenic cyst if bronchial epithelium were present, thoracic duct cyst or lymphangioma if lymphocytes were present, pleural or pericardial cyst if mesothelial cells were present, hydatid cyst if hydatid scolices were present, tuberculosis if acid-fast bacilli were present and cystic neoplasm if malignant cells were present.,11 However, the sensitivity of cyst composition analysis to determine the origin of cysts is undesirable, a study by Aravena C showed that only 27% of cysts were able to identify the origin by an analysis of the cysts composition. Nevertheless, from a therapeutic point of view, this does not affect the application of EBUS-TBNA in mediastinal cystic lesions, as patients usually presented with symptoms of compression of adjacent organs by oversized cysts or infection, independent of the cyst origin.For cysts that are difficult to diagnose on imaging , surgical excision of the cyst is of definite significance for the diagnosis and identification of the lesions origin. However, for patients who are unfit or reluctant for surgery, transcutaneous, transesophageal or transbronchial aspiration of the cystic fluid and analysis of its cellular composition can help to identify the origin of the cyst. For symptomatic mediastinal cysts, it is agreed that surgical removal of the cyst is a more reasonable option, while the choice of treatment for asymptomatic patients still remains controversial. Other than surgery, aspiration of the cyst is still an option, and it can be performed transcutaneous, transesophageal or transbronchial according to the location. Transbronchial aspiration seems to be a feasible option, as most of the mediastinal cysts are located near the trachea or bronchi. With the widespread use of EBUS-TBNA in clinical practice, the position of the needle tip can be monitered under ultrasound guidance and the depth of puncture can be adjusted, which is conducive to adequate drainage of the cystic fluid, especially for multifocal cysts. Zhong and colleagues reported that EBUS-TBNA has significant therapeutic effects on mediastinal lymphangioma without serious complications. The disease was stable over a period of 9 months to 2 years. However, another study reported an efficacy of no more than 46% after aspiration with EBUS-TBNA, with only 5.5% of patients achieving complete remission. The use of EBUS-TBNA has related complications such as secondary infection and hemorrhage, especially for cystic lesions, which seems to have a greater potential to develop cystic infection, lung abscess, pleural infection or mediastinal infection than solid lesions of the mediastinum.,13,16,25 The reported infectious pathogens include Streptococcus,–25,30–31 Coryneform bacteria and Prevotella melaninogenica, Staphylococcus,,24 Mycobacterium tuberculosis, Haemophilus Influenza, Eikenella corrodens,,33 Klebsiella pneumonia, Actinomyces odontolyticus, Gemella morbillorum, Prevotella buccae, Actinomyces, among which the most of the reported bacteria are oral flora. During the procedure of EBUS-BNA, the bronchoscope and needle may be contaminated when passing through the oropharynx, thus carrying bacteria of the oral flora into the cyst. Mediastinal cysts, however, have no vascular component and have a compromised ability to eliminate bacteria, and therefore are predisposed to secondary infections. Other sources of contamination include needle contamination by slides during specimen preparations, inadequate instrument sterilization.Due to the rare nature of mediastinal cysts, there is still little evidence to guide management, and clinical recommendations are mostly derived from case reports.Hence, on the basis of our experience and literature review, we suggest judicious use of EBUS-TBNA as a therapeutic tool for mediastinal cystic lesions as it might be associated with increased risk for infection. In conclusion, EBUS-TBNA is a useful diagnostic and therapeutic tool for the management of mediastinal cysts, including lymphangioma. However, considering the possibility of serious complications, the clinical procedure should be carried out scrupulously with appropriate patient selection and strict aseptic principles.The authors express their gratitude to Dr E.K. Makoni in Parirenyatwa group of Hospital in Zimbabwe and Dr Jie Shu in Xiangya Hospital, Central South University in China for their advices on language editing.Funding acquisition: Zhi-Guang Liu.Investigation: Wei-Dong Zhang, Huai-Qiu Wu, Zhi-Guang Liu.Resources: Yong-Xue Wang.Writing – original draft: Wei Liu.Writing – review and editing: Wei Liu.


Acute coronary syndrome (ACS) is prevalent in continuous ambulatory peritoneal dialysis (CAPD) patients. However, the association between the apoprotein profile and ACS is not well known. Therefore, we aimed to investigate the relationship between apoproteins and ACS in CAPD patients.Eighty-one CAPD patients were included in this retrospective study. The primary endpoint was ACS. Predictors were baseline apoprotein levels, particularly the ratio of apoprotein A1 (Apo A1)/apoprotein B (Apo B). Cox regression was used to determine the relationship between Apo A1/Apo B and ACS.p = 0.03), low-density lipoprotein cholesterol (LDL-C) (p = 0.04), C-reactive protein (p = 0.01), and Apo B (p < 0.01). However, hemoglobin (p = 0.01) and Apo A1/Apo B (p < 0.01) were lower in the ACS group than the non-ACS group. Patients with Apo A1/Apo B ≥ 1.105 experienced fewer ACS compared with those with Apo A1/Apo B < 1.105 . In Cox regression, Apo A1/Apo B  was independently associated with ACS.During follow-up, 34 (41.98%) CAPD patients experienced an ACS. ACS patients had higher levels of total cholesterol (Apo A1/Apo B was strongly associated with ACS and may be considered as a predictor of future ACS in CAPD patients. More than 10% of people worldwide have chronic kidney disease (CKD) ,3. AlthoHyperlipidemia is a recognized risk factor for CVD in the general population . Dyslipin = 11), missing data (n = 19), had peritoneal dialysis failure during follow-up (n = 5), and lost to follow-up (n = 4) were excluded. The study was carried out following the Declaration of Helsinki and approved by the Medical Ethics Committee of the Third Affiliated Hospital of Soochow University (Number: 2014-24).As shown in From enrollment, data on age, gender, body mass index (BMI), smoking, hypertension, diabetes, history of CVD, primary disease of CKD, use of statins, residual glomerular filtration rate (GRF), serum creatinine, urea nitrogen, lipid profile, albumin, hemoglobin, intact parathyroid hormone (iPTH), serum calcium, serum phosphate and C-reactive protein (CRP) were collected from case records. The above laboratory parameters were measured by the central laboratory in an automatic system and a standardized method. The Apo A1/Apo B ratio was obtained by dividing the absolute value of Apo A1 by the absolute value of Apo B.The enrolled CAPD patients were followed up by outpatient, inpatient, telephone, or home visited. The primary endpoint of this study was the occurrence of ACS during follow-up. ACS includes acute ST-segment elevation myocardial infarction (STEMI), acute non-ST-segment elevation myocardial infarction, and unstable angina pectoris, which were diagnosed by clinical symptoms, dynamic electrocardiogram changes, and serum myocardial injury markers.v. 19.0, IBM Corp., Armonk, NY). Continuous variables were expressed as mean ± standard deviation, and categorical variables were presented as frequency and percentage. The cutoff values of Apo A1/Apo B were analyzed using ROC curves. Survival curves according to Apo A1/Apo B were determined by K-M survival analysis. Cox regression analysis was performed to evaluate the association between Apo A1/Apo B and ACS. A 2-tailed p < 0.05 was considered statistically significant.Statistical analysis was performed using SPSS software (p = 0.76), sex (p = 0.50), BMI (p = 0.86), smoking (p = 0.38), hypertension (p = 0.81), diabetes (p = 0.79), history of CVD (p = 0.51), etiology of CKD (p > 0.05), use of statins (p = 0.44), Follow-up duration (p = 0.44), total t/vureak (p = 0.07), residual GFR (p = 0.55), urea nitrogen (p = 0.17), serum creatinine (p = 0.43), triglycerides (p = 0.11), high-density lipoprotein cholesterol (HDL-C) (p = 0.22), albumin (p = 0.65), iPTH (p = 0.97), serum calcium (p = 0.87), serum phosphate (p = 0.80), and Apo A1 (p = 0.09) between the CVD and non-CVD groups. Moreover, as showed in p = 0.03), low-density lipoprotein cholesterol (LDL-C) (p = 0.04), CRP (p = 0.01), and Apo B (p < 0.01). However, hemoglobin (p = 0.01) and Apo A1/Apo B (p < 0.01) were significantly lower in the ACS group than in the non-ACS group.Eighty-one CAPD patients were included in this retrospective study. During follow-up, a total of 34 (41.98%) patients experienced ACS. Patients were divided into two groups according to the presence or absence of ACS. No statistical differences were found in age .In the second step, the optimal cutoff value of Apo A1/Apo B was calculated using ROC curve analysis. As shown in p < 0.01). In Cox regression (p = 0.03) was independently associated with ACS in CAPD patients, Even after adjustment for age , history of CVD , diabetes , smoking , triglycerides , total cholesterol , CRP , hemoglobin , albumin  and Apo B .According to gression , Apo A1/r = 0.03, p = 0.77), Apo B , or Apo A1/Apo B .Finally, bivariate correlation analysis was used to evaluate the association between apoproteins and CRP. As a result, no correlation was found between CRP and Apo A1  and is closely related to CVD. A study  of 299 pp < 0.01). This suggests that Apo A1/Apo B may be well used as a predictor for the future occurrence of ACS in CAPD patients.The Apo A1/Apo B ratio provides a better summary of the burden of dyslipidemia than conventional lipids and lipoproteins. Some studies have confirmed that the Apo B/Apo A1 ratio was superior to LDL-C and HDL-C in predicting cardiovascular events. Bodde et al.  found thAndronesi et al.  found thInflammation may affect Apo A1 metabolism to some extent in patients with renal failure, and CRP was found to be significantly higher in the CVD group than in the non-CVD group . Our stuOur study has some limitations. This study was a retrospective small sample study, and the conclusion needs to be confirmed by prospective large sample studies in the future. Secondly, due to the voluntary principle, the enrolled patients pay more attention to their physical health than the non-enrolled patients, so the research population may have a certain degree of selection bias, which cannot well represent the general CAPD patients. Thirdly, some characteristics, such as low BMI and low incidence of diabetes/diabetic nephropathy, were unique to this study population, i.e. Asian CKD patients. This limits the generalizability of the data. Finally, it was unclear how dyslipidemia affected the development of ACS.In conclusion, we found that the Apo A1/Apo B ratio was independently associated with ACS in CAPD patients. Our study suggested that the ratio of Apo A1/Apo B may be considered as a predictor of future ACS risk compared with traditional lipid measures. The role of apolipoproteins in predicting ACS in CKD patients and their specific mechanisms require further investigation.


It was suggested that to resolve the difference in these outcomes among the prefectures, being obese or underweight needs to be addressed in mothers.The differences in the rates and trends of the overall low birth weight and term low birth weight in recent years are unknown for the Japanese prefectures. In this ecological study, we revealed the rates for each prefecture and investigated the factors affecting the regional differences in these outcomes. Aggregated vital statistics data from 2007 to 2019 were obtained from the Ministry of Health, Labour, and Welfare in Japan. The association between the outcomes and the variables, including the infants’ birth characteristics, medical characteristics, and socioeconomic characteristics of the prefectures, were analyzed. An analysis of repeated-measures data was conducted using the data from 2013 and 2018 for each prefecture. The trend for the rates of overall low birth weight and term low birth weight over the years differed among the prefectures. Moreover, the proportions of multiple births and lean (body mass index <18.5 kg/m The rates of low birth weight are representative of perinatal health outcomes worldwide and are one of the indicators of maternal and child health. Low birth weight is associated with preterm birth and intrauterine growth retardation. Low birth weight infants have a higher risk of neonatal and infant mortality or cardiovascular diseases ,2,3,4; tRegional differences in the rates of preterm birth or low birth weight within a country have been investigated in other countries, and a disparity in the outcomes has been observed ,11,12,13In this study, we assessed the rate of low birth weight in prefectures in Japan and investigated the factors affecting these differences by an ecological study using all of the birth data in the vital statistics.Vital statistics data for births were used for the analysis. We requested the aggregated data from the Ministry of Health, Labour, and Welfare by a made-to-order aggregation based on the Statistics Act . Data on the number of births by maternal age group, year, prefecture, birth weight, and gestational age were obtained. Data from 2007 to 2019 on all 47 prefectures in Japan were obtained. In this study, we used the operational definitions for the two outcomes, i.e., low birth weight rate and the term low birth weight. Low birth weight was defined as infants whose birth weight was less than 2500 g ,18. The 2) and of obese persons (BMI ≥ 25.0 kg/m2) was calculated. Although obesity is often judged based on the cut-off value of BMI ≥ 30.0 kg/m2 across the world [2 is generally used in Japan and Korea [To investigate an association between the outcomes and predictors via an ecological study using prefectural data, we used variables related to the medical characteristics, birth characteristics, maternal characteristics, and socioeconomic factors of the prefectures: the proportion of multiple births (%); the proportion of live births from mothers aged under 20 years (%); the proportion of live births from mothers aged 40 years or more (%); the number of hospitals per 100,000 persons; the number of clinics per 100,000 persons; the number of hospitals with an obstetrics department per 100,000 persons; total population; population density; taxable income per capita (1000 yen); the proportion of female high school graduates going to a higher educational institution; the proportion of live births from unemployed households; the proportion of lean women (%); and the proportion of obese women (%) for each prefecture. Data on the number of multiple births and total births for each prefecture were obtained from the vital statistics . Taxablehe world ,24, the ns BMI ≥ .0 kg/m2 nd Korea ,27,28,29https://cran.r-project.org/bin/windows/base/old/3.6.3/, accessed on 23 February 2022).For the statistical analysis, we calculated the overall low birth weight rate and term low birth weight rate for each prefecture by year. In addition, we calculated the overall low birth weight rate and term low birth weight rate in the analyzed periods for each prefecture and calculated the annual percent change (APC) of the outcomes for each prefecture. The APC was calculated by applying a linear regression model to the logarithm of the rate using year as an explanatory variable. Moreover, an ecological study was conducted to investigate an association between the outcomes and the characteristics of the prefecture. An ecological study is a type of epidemiological study that is conducted to investigate an association between a health outcome and risk factors using regional-level data as opposed to individual-level data, and it aims at identifying the risk factors of the health outcome by focusing on the differences in the regional characteristics. Therefore, regression analysis was performed using the data of each prefecture in this study. An analysis using the repeated-measures data of the prefectures was conducted using the data from 2013 and 2018 in each prefecture. A linear mixed effects model was used using the logarithm of the rate as an outcome, and each prefecture was treated as a random effect in the analysis. When regression analysis is used for data where multiple observations exist for a subject , a linear mixed model is often used. All the explanatory variables were scaled, and a standardized partial regression coefficient (SPRC) and its 95% confidence interval were calculated for each variable to compare their effects on the outcomes. The SPRC is a type of regression coefficient that is obtained when an explanatory variable is scaled before applying the regression model. By calculating the SPRC, we could compare the degree of association of an outcome variable among the explanatory variables. A statistically significant difference was set as <0.05, and all the statistical tests were 2-sided. All the statistical analyses were conducted using R version 3.6.3 (2) and obese women (BMI ≥ 25 kg/m2) had a statistically significant positive association with both the overall low birth weight rate and the term low birth weight rate. Therefore, the proportions of lean and obese women in a prefecture were positively associated with the prefectural low birth weight rate. In addition, the absolute value of SPRC for the proportion of obese women was the largest among the explanatory variables, and the proportion of obese women was suggested to be the largest factor related to the regional differences in low birth weight.In this ecological study, we aimed to reveal the low birth weight rates in different prefectures in Japan and identify the factors that determine the differences in these rates. As a result, we revealed a trend in the low birth weight rate for each prefecture in Japan and identified some factors associated with these regional differences. The proportion of multiple births were statistically significantly associated with low birth weight and term low birth weight, and the proportion of births by women with advanced age was statistically significantly associated with the term low birth weight rate. Multiple births are a well-known factor contributing to preterm birth and low birth weight . In addiAs shown in the results, the number of births in Japan is rapidly decreasing over the years. Tendencies to not marry and marry late are directly related to the decline in fertility in Japan, and factors such as the increased burden of parenting, women’s empowerment, and an increase in the rate of irregular employment are believed to be the possible causes . In addiPre-pregnancy obesity and underweight are both risk factors for low birth weight in infants ,38. In pIt has been suggested that regional differences in obesity and underweight possibly affect the regional differences in low birth weight among prefectures. Preventive measures for obesity are often promoted to prevent lifestyle-related diseases; however, obesity is also related to the birth weight of infants. Current differences in health statuses among prefectures can lead to differences in health statuses in subsequent generations. It has been suggested that lower maternal birth weight is associated with low-birth-weight infants in Japan . TherefoDisparities are known to exist in the rates of both infant and adult mortality among prefectures in Japan ,50, and We revealed regional differences in the rates of low birth weight and term low birth weight, and identified characteristics of prefectures associated with the regional differences by an ecological study using all of the birth data in Japan. As a result, we can see that trends in the rates differed depending on the prefectures in the analyzed periods. As a result of this ecological study, the proportion of live births from mothers aged 40 years or more had a statistically significant positive association with the rate of term low birth weight. In addition, the proportion of multiple births and the proportions of lean and obese women had a statistically significant positive association with both the low birth weight rate and term low birth weight rate.


Trophoblast inclusions—cross sections of abnormal trophoblast bilayer infoldings—have previously been associated with aneuploidy, placenta accreta, and prematurity. This study was conducted to establish the relationship between trophoblast inclusions and a range of placental, pregnancy, and birth outcomes in a patient population with high smoking and alcohol exposure. Specifically, we sought to evaluate the association between the presence of trophoblast inclusions and 1) three primary birth outcomes: full-term birth, preterm birth, and stillbirth; 2) gestational age at delivery; and 3) specific placental pathologies.Two slides containing chorionic villi were evaluated from 589 placentas that were collected from Stellenbosch University in Cape Town, South Africa as part of the prospective, multicenter cohort Safe Passage Study of the Prenatal Alcohol and SIDS and Stillbirth Network. The subsample included 307 full-term live births, 212 preterm live births, and 70 stillbirths.p = 0.002), with an even greater odds ratio for placentas from stillborn compared to term liveborn deliveries . Gestational age at delivery was inversely associated with trophoblast inclusion frequency. Trophoblast inclusions were significantly associated with small for gestational age birthweight, induction of labor, villous edema, placental infarction, and inflammation of the chorionic plate.We found that the odds of identifying at least one trophoblast inclusion across two slides of chorionic villi was significantly higher for placentas from preterm compared to term liveborn deliveries (OR = 1.74; 95% CI: 1.22, 2.49, The novel associations that we report warrant further investigation in order to understand the complex network of biological mechanisms through which the factors that lead to trophoblast inclusions may influence or reflect the trajectory and health of a pregnancy. Ultimately, this line of research may provide critical insights that could inform both clinical and research applications. Throughout pregnancy, the placenta serves as the critical interface between mother and fetus. Instances of abnormal development and functioning of the placenta have been implicated in maternal and fetal mortality and numerous morbidities including preeclampsia, maternal hemorrhage, preterm birth, and intrauterine growth restriction ,2. ThereTypical placental development involves the fusion of mononuclear cytotrophoblast cells into the multinucleated syncytiotrophoblast layer . When eiTo gain deeper insight into the relationship between TIs and pregnancy, birth and placental pathologies, we performed a rigorous histologic examination on a subset of n = 589 placentas from the Safe Passage Study (SPS) of the Prenatal Alcohol and SIDS and Stillbirth (PASS) Network ,12 to asThe SPS, a prospective, multicenter cohort study, enrolled n = 10,088 women  during pregnancy between August 2007-October 2016 across multiple study sites in North Dakota and South Dakota in the United States, and in Cape Town, South Africa ,12. The Pregnant women were eligible to participate if they 1) were able to provide informed written consent; 2) were pregnant with one or two fetuses; 3) were ≥16 years of age; 4) had a gestational age (GA) of at least 6 weeks + 0 days and were not at the delivery admission unit; and 5) were fluent in English or Afrikaans.To accommodate simultaneous sub-analyses of the data, slides were requested and retrieved for n = 594 placentas based on one or both of the following criteria: 1) the pregnancy resulted in a preterm live birth or a stillbirth between 20 weeks + 0 days and 43 weeks GA (n = 282) or 2) the pregnancy resulted in a full-term live birth and the infant subsequently participated in at least one follow-up assessment between birth and 4 years of age (n = 307). Of the 594 births included in this sub-analysis, there were 576 singleton pregnancies and 9 sets of twins.Tissue was collected from 28% of the participants as part of an embedded study . Whole pThe two de-identified slides from each placenta were evaluated at the Yale Reproductive and Placental Research Unit. The 1,188 slides were randomized and re-labeled in numeric order and were blindly read by a placental pathologist (HJK). The pathologist was blind to all demographic, clinical, prenatal, birth outcome, and developmental outcome data. All slides were examined to determine the frequency of TIs. Based on previous methods, we determined the total number of inclusions and calcified inclusions ,8,10. UsPregnancy and birth outcomes were characterized in detail as part of the SPS study ,12. Medith percentile), appropriate for GA (10th-90th percentile), or large for GA (>90th percentile). As part of our pathologic examination, we recorded the absence or presence of villous edema  [Birth characteristics and measures of placental pathology were determined using data obtained as part of the original SPS study and through histological examination of the slides. From the original study records, we obtained the mode of delivery , labor onset (spontaneous versus induced), GA at delivery , and birth weight relative to GA. Using the trimmed weights for all placentas collected through the SPS study n = 1556), we generated growth curves of placental weight for GA at delivery separately for males and females . Placentus core) , the abs56, we geus core) , the absus core) , increasus core) , or the us core) .We used descriptive statistics, mean and standard deviation for continuous variables, count and percentages for categorical variables to describe the characteristics of the study population by birth outcome groups. We used logistic regression models to estimate the odds ratio and 95% confidence intervals of presence of TI associated with birth outcomes. Separate models were fit for each birth outcome. Models were adjusted for maternal age at delivery, education status, marital status, parity, maternal height, and prenatal depression as potential confounders. A missing indicator method was used in multivariate models when a covariate was missing. To determine whether the association between TIs and birth outcome was dependent on GA at delivery, we compared placentas from livebirths to those from stillbirths and performed a linear regression model that included GA at delivery as a covariate. We used separate Univariate logistic regression models to examine the association of TIs with birth characteristics and placental pathologies. All analyses were performed in SAS software version 9.4  and Python packages of statsmodels and Matplotlib.The study sample included n = 307 FTLBs, n = 212 PTLBs, and n = 70 SBs . Demographic and clinical characteristics for the overall sample and each birth outcome group are presented in p = 0.002), while the odds of SB compared to FTLB placentas was 4.95 . After adjusting for maternal age at delivery, education status, marital status, parity, maternal height, and maternal depression, the adjusted odds ratios for PTLB and SB compared to FTLB placentas were 1.71  and 4.14 , respectively (p = 0.0002).Of the 589 placentas examined, at least one TI was observed across the two slides from 108 (35%) of the FTLB placentas, in 103 (49%) of the PTLB placentas, and in 51 (73%) of the SB placentas. Compared to the FTLB placentas, both PTLB placentas and SB placentas were significantly more likely to have TIs. Specifically, the odds of identifying one or more TIs in two slides from PTLB compared to FTLB placentas was 1.74  (p<0.001) and 4.24 weeks earlier than those with ‘mild’ TIs (p<0.001).Infants with no TIs (‘none’) were delivered between 21.9–42.4 weeks GA, with a median of 38.3 weeks GA and a mean of 37.1 (±3.68) weeks GA at delivery. Infants with ‘mild’ TIs (an average of >0–5 TIs per slide) were delivered between 20.9–43.0 weeks GA, with a median of 36.6 weeks GA and a mean of 35.0 (±5.44) weeks GA at delivery. Infants with ‘marked’ TIs (an average of >5 TIs per slide) were delivered between 20.9–40.0 weeks GA, with a median of 32.4 weeks GA and a mean of 30.8 (±6.63) weeks GA at delivery. To determine if there was a dose-dependent effect of TIs on GA at delivery, we performed an adjusted linear regression. Compared to infants with no TIs, infants categorized as having ‘mild’ TIs were delivered 2.1 weeks earlier ( <0.001) . Consistp’s >0.05). TIs were also not statistically significantly associated with placental weight percentile or any of the other pathologic features examined for this study, regardless of the inclusion of GA in the models (p’s >0.05).The presence of villous edema, placental infarction, maternal neutrophilic infiltration of the chorionic plate (plate inflammation), spontaneous labor onset, and neonatal weight that was small for GA were all associated with a statistically significantly higher odds of observing TIs . HoweverWe report a critical replication and expansion of the previous literature on the role of TIs in placental, pregnancy, and fetal health in a cohort of women with high smoking and alcohol intake exposure. Consistent with an earlier finding , the oddIn a previous study, we did not find an association between TIs and clinical evidence of preeclampsia, gestational diabetes, or hypertension . HoweverThat none of the other placental pathologic features were associated with TIs, regardless of the inclusion of GA at delivery as a covariate, suggests that the etiology of TIs is independent of the majority of pathologic processes that can impact the placenta. This is consistent with our overarching hypothesis that TIs are a biological marker of genetic and/or developmental abnormality and is consistent with the many studies that have identified TIs in both aneuploid –30 and ein-utero factors that may contribute to the incidence of TIs, this relationship may inform clinical management of subsequent pregnancies that are preceded by those in which TIs have been identified.Our analysis has established an association between TIs and several clinically-relevant outcome measures including preterm birth, stillbirth, small birth weight for gestational age, and specific placental pathologies. Both prematurity and stillbirth are associated with an increased odds ratio of identifying TIs, providing insight into a potential biological process underlying these adverse birth outcomes. Further, this is the first analysis to directly evaluate the relationship between the quantity of TIs and gestational age at delivery. We found that placentas deemed most severe with regards to TI frequency were delivered at a significantly earlier GA than those with no or mild TI frequencies. Given both genetic and The associations reported here support the biological plausibility that TIs may indicate dysfunction and/or may play an important role in shaping pregnancy and fetal development. Our data indicate that greater TI severity is associated with lower GA at delivery. However, the observed increased prevalence of TIs in placentas delivered earlier in gestation does not necessitate the same prevalence of TIs in normal undelivered placentas at these same early gestational ages. Earlier literature ,27–30,36Given the robust relationships between TIs and the pregnancy, placental, and birth outcomes that we analyzed, future research aimed at understanding the biological mechanisms underlying such outcomes should consider including TIs in the mechanistic pathway. Additionally, further research is needed to determine potential environmental and health-related factors (e.g. tobacco and ethanol exposure), or even placental pathologies that may account for the greater-than-expected proportion of full-term placentas with at least one TI (35% versus previously reported prevalence of 2–8% ,9,10 notThe main strength of our study was the relatively large sample size which was specifically collected for research purposes, rather than relying on information obtained solely for clinical purposes. Therefore, we were able to generate a very detailed and deeply-phenotyped dataset that allowed us to include many potential confounding variables. In a prior publication on the same dataset, we established very high test-retest reliability of the pathologist’s TI counts . This daOur initial sample selection was blind to all information regarding the pregnancy and birth outcomes, with the exception of GA at delivery regardless of whether the infant was liveborn or not. Thus, it was not until unblinding of the birth outcomes that we discovered how many stillbirths had been captured in our subsample. Similarly, the sample was not randomly selected based on birth outcome, but rather, was generated based on the specific inclusion criteria previously described. This selection process may have involved unintended bias in sample selection. To address these limitations, it would be beneficial to conduct a prospective analysis of TIs with long-term birth and developmental follow-up. An additional limitation is that we did not have full access to the gross examination of the placenta and rather, were restricted to analyses of two randomly selected slides. This limitation may have impacted our ability to characterize specific features such as the number and location of placental infarcts. Further, genetic analyses were not performed on placentas included in these analyses, and therefore, our data cannot address the biological linkage between genetics, including mosaicism, and the development of TIs. A final limitation of this study is that all slides were reviewed by a single pathologist at one institution, however, it is notable that the pathologist had very high test-re-test reliability .This paper provides foundational associative information regarding the potential role of TIs in a range of placental, pregnancy, and birth processes. The novel associations that we report warrant further investigation in order to understand the complex network of biological mechanisms through which TIs may influence the trajectory and health of a pregnancy. Ultimately, this line of research may provide critical insight that will inform both clinical and research applications.


Similarly, the Enterobacteriaceae counts of candies from formal processors (nondetectable) were considerably lower (p = 0.015) than those from informal processors (1.8 ± 0.56 log 10 CFU/g). The ergosterol content in these baobab product samples ranged between 0.46 and 1.92 mg/100 g while the aflatoxin content ranged between 3.93 and 11.09 × 103 μg/kg, respectively. Fungal and aflatoxin contamination was detected in 25% and 5% of pulp from formal and informal processors, respectively, and in 5% of candies from informal processors. Microbial contamination in processed baobab products shows an unhygienic processing environment while the fungal and aflatoxin contamination may indicate poor postharvest handling, transport, and storage conditions of baobab fruits along the baobab value chain.Baobab fruit demand has been on the rise in the recent past, and in an attempt to match the demand, farmers and middlemen are forced to harvest immature fruits which are not fully dried. To ensure an acceptable moisture content, baobab fruits are subjected to solar drying, which is a slow process and often carried out in open and unhygienic conditions raising safety concerns. This study was conducted to investigate the microbial and aflatoxin contamination levels in ready-to-eat baobab products from selected formal and informal processors in specific counties of Kenya. Selected processed baobab products were sampled randomly from formal and informal processors and analyzed for the total aerobic count, Enterobacteriaceae, yeast and molds, ergosterol, aflatoxins, moisture, and water activity. The moisture and water activity of baobab pulp and candies from formal processors ranged between 7.73% and 15.06% and 0.532 and 0.740 compared to those from informal processors which ranged from 10.50% to 23.47% and 0.532 to 0.751, respectively. In this study, baobab pulp from formal processors had significantly ( Adansonia digitata) candies and pulp are ready-to-eat snacks processed from baobab fruit mostly consumed in a number of sub-Saharan Africa countries. The baobab fruit pulp is naturally dried and is rich in vitamin C, calcium, and antioxidants [Baobab . HoweveSome of the handling practices of baobab pulp may also expose the pulp to extrinsic environmental factors such as high humidity. The pulp is hygroscopic and is thus susceptible to moisture absorption leading to mold growth and subsequent spoilage and/or contamination by mycotoxins rendering the commodity unsafe for consumption . DetermiAspergillus parasiticus and Aspergillus. flavus L. strains were recovered from baobab fruit obtained from the market in Zambia, though the average aflatoxin concentrations in the baobab were below maximum allowable levels in food [ in food . HoweverAdansonia digitata L.) products in eastern and coastal Kenya” [p′ = proportion ratio, x = the size of the proportion, n = the size of the total population.This study involved the use of two classification groups: the formal and the informal processors. The term formal was used in this study to indicate processors registered with the Kenya Bureau of Standards (KEBS), whereas informal represents processors not registered with KEBS. Total population sampling was applied for formal processors (control) due to the manageable size of the population and systematic sampling for the informal processors. Details of the formal baobab product processors were obtained from KEBS where a sampling list of 15 processors was obtained with only 10 of the processors selected for this study being fully operational. For formal processors, eight processors were located in Nairobi, one in Kilifi, and one in Mombasa counties. The details of the informal processors were gathered from a baobab processor survey carried out under the project “A value chain analysis of baobab  and moisture content measurements. The water activity was measured as described in, using HygroPalm HP23-AW-A Portable Water Activity Analyzer . The water activity meter was set at awQuick mode, and sealed samples were kept at relatively constant temperature (22 ± 3°C) for 5 minutes to allow for the temperature conditions of the sample and the probe to stabilize before the displayed aw reading was recorded.All samples were subjected to water activity  were made and analyzed for [The baobab samples were each analyzed in duplicates for the total aerobic count, Enterobacteriaceae, and yeast and molds according to the AOAC microbiological method number ISO 4833, ISO 21528-2, and ISO 7954, respectively . All medyzed for : total ayzed for , 18. Ther2) of 0.999 was used to demonstrate linearity. Based on the signal-to-noise ratios (S/N) of 3 : 1 and 10 : 1, respectively, the limit of detection (LOD) and limit of quantification (LOQ) were determined [μg/ml in acetonitrile solution. Working standards were prepared by serial diluting with acetonitrile, and a 6-point calibration curve was generated covering the range of 1.25–12.5 ppm for each aflatoxin group.Ergosterol standard (E6510-5G) was obtained from Sigma-Aldrich . The ergosterol stock solution was prepared at a concentration of 1 mg/ml, and further working standards were prepared by serial dilutions with ethanol. A standard calibration curve was then generated, and a correlation coefficient of . The HPLC system consisted of an auto-sampler model SIL-20A HT, quaternary pumps (Shimadzu model LC-20 AD), a reverse-phase SUPELCO C-18 (5 μm 260 mm × 4.6 mm) column, CTO-10AVP column oven, and a photodiode detector (SPD-M20A). The separation was achieved through isocratic elution at a flow rate of 2 ml/min, with a linear gradient solvent .Extraction and quantification of ergosterol were carried out on each of the 60 baobab samples. Two grams of the sample was placed in the saponification solution composed of 15 ml of methanol and 5 ml of potassium hydroxide solution (40 g/l KOH in ethanol) and agitated for 30 minutes. This was then filtered into evaporation tubes using a 0.2 nitrogen . The ergv/v) and centrifuged at 10,000 rpm for 5 minutes. Five millilitres of supernatant was transferred to a 50 ml centrifuge tube and diluted with 40 ml of 2% PBST  solution [μm millipore filters into vials. Aflatoxin quantification was done by injecting 20 μl of the samples into a SHIMADZU HPLC system  equipped with; RF20A fluorescence detector operated at an excitation wavelength of 350 nm and an emission wavelength of 450 nm; auto-sampler SIL 20AHT, a reverse-phase SUPELCO C-18 (5 μm 260 mm × 4.6 mm) column, CTO 10ASVP column oven set at 40°C, and LC20AD quaternary pump. A mobile phase of acetonitrile : methanol : water (10 : 30 : 60 v/v) was used at a flow rate of 1 ml/min.Two grams of the baobab samples was added to an extraction solution comprising of 1 g sodium chloride in 25 ml  than those from formal processors   . This cop ≤ 0.001, p ≤ 0.001) Enterobacteriaceae and yeast and molds counts (3.1 ± 0.70 log10 CFU/g and 5.3 ± 0.11 log10 CFU/g) than those from formal processors (0.7 ± 0.29 log10 CFU/g and 3.1 ± 0.38 log10 CFU/g), respectively. Production practices, extrinsic, intrinsic, and processing factors determine the microbial load of food products at the time of consumption [The presence of microbes in food does not imply unfitness for consumption, but rather an indication of the hygienic status during preparation and processing . Nonethesumption .The total aerobic count loads in this study (4.3 ± 0.22) in pulp were higher than those of studies done on microbial analysis of freeze-dried fruits  and the The baobab pulp contamination can be traced back to handling practices along the value chain and at the farm level . To meetp = 0.015) was observed in terms of Enterobacteriaceae counts between candies from formal (0.0 ± 0.00 log10 CFU/g) and informal processors (1.8 ± 0.56 log10 CFU/g). The Enterobacteriaceae counts in baobab products in this study were higher than those reported by Ntuli et al. [10 CFU-g)   in some 0 CFU-g) . During 10 CFU/g) in the candies from the informal processors  in the total aerobic count, yeast and molds, and Enterobacteriaceae loads, respectively, among products from the regions. The highest TAC contamination in candies was detected in samples collected from Mombasa County (5.4 ± 0.28 log10 CFU/g), whereas the highest Enterobacteriaceae contamination was detected in samples collected from Kilifi county (3.3 ± 0.11 log10 CFU/g)  . This co et al., ).10 CFU/g and 5.1 ± 0.01 log10 CFU/g, respectively. Similarly, candies from informal processors in Mombasa and Kilifi recorded the highest levels of yeast and molds at 4.3 ± 0.98 log10 CFU/g and 3.6 ± 0.05 log10 CFU/g, respectively, as shown in (10 CFU/g) and those from Mombasa and Kilifi (5.4 ± 0.30 log10 CFU/g and 5.1 ± 0.01 log10 CFU/g), respectively. A similar trend can be seen in a study done on the effect of temperature on microbial growth in food during storage, where high environmental temperatures increase the proliferation of microorganisms [The baobab pulp from informal processors in Mombasa and Kilifi recorded the highest yeast and mold contamination in baobab pulp at 5.4 ± 0.30 logshown in . A signirganisms . Kilifi rganisms ).μg/kg for AFB1 and 10 μg/kg for total aflatoxins. The EU standard for baobab pulp prescribes a maximum limit of 2 μg/kg for AFB1 and 4 μg/kg total aflatoxins. Temperature, water activity, and humidity play a vital role in fungal growth and mycotoxin production [r = 0.5019, p = 0.04) , , and , respectively. The results obtained are comparable to the prediction of mycotoxin deoxynivalenol (DON) contamination in Fusarium-infected wheat grains based on the determination of the ergosterol content [oduction . A posit = 0.04) . There wSamples 1, 7, 12, and 14 had a water activity of above 0.7 , which cThe percentage of the fungal and aflatoxin-contaminated baobab samples is shown in μg/kg in the analyzed samples indicating a potential health risk to baobab consumers [The results from the study show that there is a minimal inspection of the on-shelf baobab products by the regulatory agencies. Aflatoxin concentrations were ranging between 3.93 and 11.09 × 103 onsumers . The cononsumers . Anotheronsumers .In conclusion, informal baobab processor products recorded the highest microbial contamination in terms of Enterobacteriaceae and TAC, which could be associated with unhygienic handling during processing. The results of this study incriminate the unhygienic postharvest practices along the baobab value chain as the source of contamination of the baobab pulp and candies. Therefore to ensure the safety of dried baobab fruits and baobab products along the baobab value chain, remedial steps such as training on good hygiene and good manufacturing practices as well as Hazard Analysis Critical Control Points (HACCP) should be taken .To maintain the postharvest quality of baobab fruits, right postharvest management steps should be taken. For instance, baobab drying should be carried out using certain methods, such as convective air drying and vacuum drying, in an enclosed environment. However, if solar drying is to be used, the use of indirect solar drying or enclosed solar dryers is recommended as the product is covered with a transparent cover in a cabinet that protects it from dust, insects, rains, and rodents that could otherwise lead to contamination. Appropriate primary and secondary packaging technologies should also be incorporated especially hermetic bag packages.


Length at birth is important for evaluating childhood growth and development. It is of interest in Pediatrics because of its implications for perinatal and postnatal morbidity and mortality. Predicting birth length will be useful in anticipating and managing possible complications associated with pregnancy and birth of babies with abnormal birth length.The aim was to identify easily accessible parental determinants of baby's birth length in Lagos, Nigeria, using a sample of patients attending a government hospital.Parental anthropometrics and other data were obtained from 250 couples by actual measurements, oral interviews and questionnaires. Baby's birth length was measured immediately after delivery by qualified, a well-trained obstetric nurse, and association between parental and offspring parameters were assessed.Weight gain, maternal weight, parity and mid-parental height were the significant parental explanatory variables of offspring birth length. They were the most suitable variables for a generated model for predicting babies' birth length from parental variables in the study.A model that might be useful for predicting babies' birth length from easily accessible parental variables was produced. This model may complement ultrasonographic data for predicting baby's birth length with a view to achieving better perinatal and postnatal care. Birth length is an important neonatal measurement taken at birth not only because of its association with adult height but also because of its importance in clinical medicine and quantitative genetics. Considerable attention had been focused on birth weight because of the general belief that it is the best indicator of foetal growth and birth size when compared to other neonatal parameters. Furthermore, birth weight has been identified by several researchers as one of the most important factors of quality of life in adulthood according to the fetal origin hypothesis proposed by Bakeret al.The association between birth length and several perinatal and postnatal morbidity and mortality has been the subject of many studies. There is a strong association between birth length and late onset wheeze in children of between 2 months to 4 years oldIn view of the high level of poverty, illiteracy and ignorance especially in low income countries, many expectant mothers in these regions do not attend antenatal clinics; foetal growth including foetal length could therefore not be properly monitored. By the time pregnancy is at full term and parturition is about to commence, it might be rather late to take significantly effective steps to prevent problems associated with abnormal birth length. Thus, ability to predict abnormal birth length from parental anthropometrics may enhance identification of parents at risk of having babies with abnormal birth length to enable proper education and early surveillance. This, hopefully, would make early preventive strategies possible.A good number of studies have been carried out on heritability and parental contribution to birth length in Caucasian and Asian populations, but little or no similar studies exist in Nigeria and many other low income countries of Africa. Since heritability is a population-specific attribute, the need to carry out similar studies in Nigeria cannot be overemphasized. The present study was therefore undertaken to see the possibility of developing a model for predicting offspring's birth length from simple parental variables using a sample of patients attending a government hospital in Lagos.Although assessing baby's development through ultra-scan technique is not new, its use in estimating neonatal anthropometric parameters has not being totally efficientTwo hundred and fifty couples attending the ante-natal clinic of General Hospital, Surulere, Lagos, Nigeria, were recruited for the study after successfully seeking approval from Institution's Ethical Committee. The major inclusion criterion was singleton pregnancy while the criteria for exclusion included presence of diabetes, hypertension, malnutrition, anaemia, malignancies and HIV/AIDS. Other exclusion criteria were presence of complications such as pre-eclampsia/eclampsia, antepartum hemorrhage, uterine fibroid or any other uterine or placental abnormalities. These were verified from medical records. Sufficiency of the sample size (250 family units) for the study was determined using the formula below.11α/2)2 P(1-P) ] / E2.......................(1)n [ of 0.5 with a margin of error (E) of 0.07 at 95% confidence level i.e Zα/2 = 1.96, the appropriate sample size (n) was found to be 196 couples (taking a couple as a unit). In view of this, the sample size of 250 couples used for this study was considered adequate.After explanation of the purpose and procedure to the study participants, they were requested to fill consent forms and questionnaires administered by the researchers to obtain personal data including gender, ethnicity, weight before pregnancy (pre-pregnancy weight) parity, gravidity, and age. The participants were guided in filling the questionnaire where necessary. Information about parental anthropometrics like mothers and fathers' weights, heights were obtained through actual measurements by a trained nurse, who recorded the measurements under the section of the questionnaire tagged “For Official Use Only”.In order to get the combined effect of the height of both parents on their babies' birth length and birth weight, mid-parental height and weight were derived according to the formulas:Mid-parental height = father's height + mother's height)/2 ...………… (2)Mid-parental weight = father's weight + mother's weight)/2 ...…..……. (3)Mid-parental BMI = father's BMI + mother's BMI)/2 ...…………...….…. (4)Immediately after delivery, the baby was cleaned of blood and other post-delivery fluids. The sex of the baby was determined and those with ambiguous genitalia were excluded from the study. Measurement of birth length and other neonatal parameters routinely measured after delivery e.g. birth weight, and head circumference) was completed within 30 minutes of delivery by a trained obstetric nurse. However, since birth length was the only baby's parameter of interest in this study, it was the only neonatal variable involved in the study. Thus, two nurses were involved in the study: a general nurse who took the parental anthropometrics and an obstetric nurse who took the neonatal measurements.The raw data were analyzed statistically using Microsoft Excel (2007) and IBM Statistics / Statistical Package for Social Scientists (SPSS) Statistics (Version 25) software packages. The initial analysis was to carry out descriptive statistics followed by comparison of mean± SE by the Student's t-test. Simple correlation procedure was used to generate a pairwise correlation matrix which was followed by partial correlation analysis to see whether associations seen with simple correlation analysis were influenced by intercorrelations. Based on this, dimension reduction using Principal Component Analysis (PCA) was used to remove redundant highly correlated variables to produce smaller number of uncorrelated variables which could effectively explain and predict baby's birth length. We used multiple regression analysis to generate regression models which were subjected to analysis of variance (ANOVA) for test of goodness of fit. In all cases, P<0.05 was considered significant.2 respectively. Parity shows the highest variation (51.8) while paternal height showed the least variation (coefficient of variation = 4.6).Summary statistics of the parents is given in In Many of the studied variables were intercorrelated as shown in Many of the intercorrelations revealed by simple correlation analysis vanished with partial correlation . In TablPrincipal component analysis (PCA) of the studied variables gave four components. The four components explained 70.4% of the variation observed in the data (only a 29.6% loss of detail). Among the variables that loaded into the same component, the most partially correlated variable with each of the dependent variable was selected for multiple regression analysis. Maternal weight, mid-parental height, weight gain and parity were the variables with the most significant loading in their respective components and were therefore selected for multiple regression for birth length.Results of the multiple regression analysis gave the unstandardized multiple regression model given below:Birth Length = 38.60 + 0.39(Parity) + 0.21 + 0.04 + 0.03Birth Length = 38.60 + 0.39(Parity) + 0.21 + 0.04 + 0.03 …………………………………………………………………. (5)The standardized multiple regression model is:Birth Length = 0.20 + 0.14 + 0.13(Parity) + 0.10 …………………………………………………………………………………... (6)The predictor variables considered in this study were mainly parental anthropometrics; they were of interest because they can be easily obtained through simple measurements. Non-anthropometric parameters included in the study are maternal age and parity; they were also considered in the study because of the ease of obtaining them along with the anthropometric parameters. While we considered information on parity reliable, we could not rely on age declared by parents with the same degree of confidence because it is generally much easier to know and remember parity than age. In many cases, the declared age was confirmed from birth certificates and hospital records. In cases where birth certificates and hospital records were not available, we relied on age declared in the questionnaire. Nevertheless, some of our subjects were either illiterates or semi-illiterates, and that increased our reservations for the correctness of the declared age. This might represent a limitation of the present study.2 and 25.9 kg/m2 respectively in our study were around the upper limit of the normal range of 18.5kg/m2 – 24.9kg/m2 according to EknoyanThe summary statistics of our sample was similar to those obtained in earlier studies. In our study, the birth length of male and female babies were 48.9 cm and 49.0 cm respectively, values comparable to 51.5 cm and 50.7 cm obtained by Miletic et al.Many of the significant simple correlations observed in this study disappeared with partial correlation analysis. This suggests that many of the studied variables were interdependent. There was a simple correlation between birth length and maternal height and between birth length and parity in our study. This was in agreement with the work of Miletic et al.Miletic et al.The regression model indicated that weight gain, maternal weight, parity, and mid-parental height are parental parameters that may be considered in determining baby's birth length. There is need for caution in applying this model because birth length is a multifactorial neonatal trait that depends on several genetic and environmental factors. This was implied by the low R-squared value obtained in the generated model. Nonetheless, the fact that the R-squared values was statistically significant suggests that the identified parental anthropometric attributes contribute to babies' birth length. Moreover, the generated model may not be applicable to other populations because of population stratification. Thus, researchers should work out possible models for predicting neonatal parameters in their respective populations. Since our model should be applied with caution, its utility therefore lies in complementing already existing conventional methods such as ultrasonography for better accuracy of baby's birth length prediction. This would encourage early intervention, and therefore prevent maternal and neonatal complications associated with abnormal birth length of babies.A model for estimating birth length of babies from easily accessible parental variables was produced. The model could complement other conventional methods of estimating baby's birth length. The derived model shows that during antenatal, focus should not be on the mother alone but also on the father as well since he also contributes to baby's birth length, an attribute of considerable obstetric importance.


Methodology. This was a prospective observational study carried out in a tertiary care hospital. Eighty-four infants born at gestational ages between 37 and 42 weeks and classified as term small for gestational age (SGA) and appropriate for gestational age (AGA) were included in the study. Estimation of acylated ghrelin (AG) concentrations was done in the cord blood at birth and in venous blood at the age of 3 months in all the infants. The correlation between AG concentrations and growth parameters at birth and at 3 months was studied.  The effect of ghrelin, a growth hormone (GH) secretagogue on growth of neonates, has been studied in the past, but not fully clarified. We aimed to investigate the relationship between ghrelin and growth parameters at birth and at the age of three months in healthy term infants. r = −0.34, p value 0.03), birth length, and head circumference  in SGA neonates. However, at 3 months, AG concentrations did not correlate with changes in anthropometric parameters in both the groups.  AG concentrations were significantly higher in SGA (236.16 ± 152.4 pg/ml) than AGA neonates (59.45 ± 20.95 pg/ml) at birth. Concentrations were observed to be negatively correlated with birth weight ( Cord acylated ghrelin concentrations are higher in SGA neonates, and the concentrations are inversely proportional to the birth weight. Hence, its role as a surrogate marker for intrauterine nutrition can be suggested. However, its concentrations do not correlate with anthropometric parameters in early postnatal growth, suggesting it may not have a direct role in postnatal growth. Small for gestational age (SGA) infants are defined as neonates with birth weight for gestational age below the 10th centile as per gender-specific reference population. SGA infants are at greater risk for perinatal morbidity, persistent short stature, and metabolic alterations like diabetes, obesity, and cardiovascular disorders later in life . Most ofGhrelin, an acylated 28 amino acid peptide, was first discovered in the stomach of rats and has orexigenic and GH secretagogue properties . In humaAcylated form of ghrelin was considered the active form of ghrelin  but a grBirth weight being the most sensitive indicator of neonatal adiposity can be used as a surrogate marker of intrauterine growth, and its correlation with cord ghrelin concentrations can serve as a reliable marker of intrauterine growth and subsequent postnatal catch-up growth. As the prevalence of SGA in India is higher than that in the western population, the aetiologies are different considering differences in genetic and environmental factors . With paThis was a single-centre prospective cohort study conducted in the Department of Pediatrics in a tertiary care setting in India. Ethical clearance was sought from the Hospital Ethics Committee (D. No. 1015/FM).Term neonates born to healthy mothers in the hospital between October 2017 and October 2019 were considered for inclusion in the study.pq/d2, where p is the prevalence, q is 1 − p, and d is the deviance (5-20% of p), p = 75, q = 25, and d = 20% of 75 = 15, 4 × 75 × 25/225 = 33.3 (10% for attrition and rounded off to 40 in each group).Sample size was calculated using Cochran's formula 4Eighty-four infants—40 SGA and 44 AGA—were analysed at the end of study .After informed consent was obtained from the parents, enrolled neonates were categorised as SGA with birth weight < 10th centile and AGA with birth weight between 10th and 90th centile as per the national growth standard charts . PretermSamples (5 ml) from the cord blood of neonates were collected in EDTA tubes, labelled appropriately, centrifuged, and treated with 1/10 N HCl. Thereafter, samples were stored at -70 degrees Celsius till the time of analysis. Estimation of ghrelin concentration was done by using Human Acylated Ghrelin Express ELISA Kit  using the ELISA  method. A detailed maternal history, gestational age, and neonatal examination including anthropometric parameters were recorded in neonatal case records.Naked birth weight was taken on an electronic weighing scale  with an accuracy of ±5 g. Length was measured using an infantometer  with a precision of ±0.1 cm. Head circumference was measured between the supraorbital ridge and the occipital protuberance using a nonstretchable tape. The average of three readings was recorded for all the parameters. At three months of age, anthropometric parameters were recorded by the same observer and venous samples for evaluation of acylated ghrelin concentrations were collected in the morning utilising the same procedure.t–test and categorical variables with the chi-square test. Statistical significance was set at p value < 0.05. All the statistical analyses were performed using version 20 of SPSS (2019).All the data was expressed as mean ± SD. Correlation between different parameters was determined using Pearson's correlation test. The differences in variables between groups were compared using Student's At 3 months, SGA infants had significant weight gain (3.17 ± +0.17 kg), increase in length (8.32 ± 0.89 cm) and head circumference (5.97 ± 0.51). AGA infants also demonstrated significant gains in weight (3.13 ± 0.24 kg), length (7.55 ± 0.72 cm), and head circumference (5.23 ± 0.72). SGA infants had significant increases in length and head circumference compared to AGA infants. Though there were no significant differences in weight gain between SGA and AGA infants in absolute values, the percent gain in weight was observed to be significantly higher in SGA infants as compared to AGA infants (147.4% vs. 106.82%).SGA infants had significantly higher cord AG than AGA infants (236.16 ± 152.40 vs. 59.45 ± 20.95 pg/ml) at birth and at 3 months (652.32 ± 171.24 pg/ml vs. 361.36 ± 83.18). Changes in ghrelin levels were significantly higher in SGA infants than in AGA infants (418.66 ± 132.96 vs. 301.90 ± 89.89 pg/ml).r = −0.34, p value 0.03), birth length, and head circumference  in SGA neonates (Acylated ghrelin level was negatively correlated with birth weight (neonates .However, at 3 months, AG concentration did not correlate with the changes in anthropometric parameters in both the groups.There were no significant differences in ghrelin concentrations among male and female infants of either group at birth and at 3 months .In our study, the baseline cord AG concentrations were significantly higher in the SGA group. Similar observation was put forth by other studies with higher cord ghrelin in SGA as well as preterm infants –27. HoweAt 3 months, there was a significant gain in length and head circumference in SGA neonates. There was no difference in weight gain between the two groups but the SGA group demonstrated a higher intragroup weight gain. Mean change in serum ghrelin concentrations was also observed to be significantly higher in SGA infants than in AGA infants. Our findings are in accordance with results in other similar studies –27, 29.The exact mechanism by which ghrelin affects the weight gain in early infancy in humans is unclear, but a simple explanation would be that lower ghrelin concentrations are associated with poor appetite, resulting in lower nutritional intake. The effect of ghrelin on growth during the perinatal period is different from that on adolescents and adults. During the initial 2 years of life, concentrations of ghrelin are high, with progressive decline during childhood and adolescence , 31. ThiThis study showed no significant gender differences in AG concentrations in both groups at birth and at 3 months of age. This was consistent with other studies which have found either no gender differences or higher ghrelin concentrations among SGA girls citing a low birth weight in the females as the reason , 26, 30.The major limitation of our study is that we have not measured growth hormone and insulin-like growth factor-1 concentrations which are the main mediators for the action of ghrelin. Another limitation is short-term follow-up of infants for 3 months only. As the ghrelin concentrations show a declining trend with age , a long-Cord acylated ghrelin concentrations are higher in SGA neonates, and the concentrations are inversely proportional to the birth weight. Hence, its role as a surrogate marker for intrauterine nutrition can be suggested. However, its concentrations do not correlate with anthropometric parameters in early postnatal growth suggesting it may not have a direct role in postnatal growth. Further research with a longer follow-up is needed to explore the role of ghrelin and other growth hormone secretagogues in catch-up growth of SGA infants.