The ectopic ciliated cyst is a rare non-odontogenic cyst which occurs as a delayed complication after maxillary sinus radical surgery; this lesion emerges due to the destruction of the sinus mucosa during the surgery and entrapment of the respiratory epithelium in the region. This lesion has been observed in very rare cases following genioplasty and bimaxillary orthognathic surgery.We reported a case of the ectopic ciliated cyst in in the mandible of a 37-year-old Iranian woman following genioplasty and Lefort 1 surgery after 2 years. Its treatment was enucleation.Long-term follow-up after cosmetic surgery of both jaws is recommended due to the probability of this cyst.  The ectopic ciliated cyst, which is also known as the post operative maxillary cyst and implantation cyst, comprises about 1.5 % of total oral cysts and is often developed as a delayed complication 6 to 49 years) after maxillary sinus surgery as a result of respiratory epithelium getting trapped in the surgical site . Most of9 years aThis is the 11th report, in English papers, of an ectopic ciliated cyst of the mandible following genioplasty and Lefort I after 2 years of surgery in the upper and lower mandible. The patient was a 37-year-old Iranian woman who had visited the surgery unit of the dentistry department due to mild inflammation in the anterior mandible area. She did not mention any systematic diseases or the use of special drugs but reported Menu Max and genioplasty (advancement of maxilla) and Lefort 1 surgery for cosmetic reasons 2 years ago . Upon the panoramic radiography view, a unilocular radiolucency with definite borders was seen in the anterior mandible area which was developed from the second right tooth to the second left tooth .Upon CT scan, the lesion had led to the perforation of the mandibular buccal bone in the anterior area .Considering the clinical signs and symptoms and radiographic images, differential diagnoses included traumatic bone cyst, glandular odontogenic cyst, odontogenic keratocyst, and central giant cell granuloma. The sample was sent to the pathology laboratory in formalin 10%.Considering the clinical signs and symptoms and radiographic images, differential diagnoses included traumatic bone cyst, glandular odontogenic cyst, odontogenic keratocyst, and central giant cell granuloma. The sample was sent to the pathology laboratory in formalin 10%.Upon macroscopic investigation, a piece of plate shaped soft gray cream tissue with elastic consistency measuring 0.3 × 0.5 × 0.7 cm in size was seen. Cross-sectional cutting of the cyst showed that the cross section was non-uniform in section. A cystic section with a non-uniform thickness of at least 1 mm was observed.Upon microscopic investigation, a cystic stratified structure with a ciliated respiratory epithelium or sometimes stratified squamous epithelium with mucous cells lacking rete peg Ectopic Ciliated Cyst in the Mandible were observed. The underlying connective tissue was fibrous and collagenized and had delicate blood vessels and few inflammatory cells. Based on pathological findings and location of the lesion, a diagnosis of ectopic ciliated cyst was made .The ectopic ciliated cyst is a non-odontogenic and common cyst in the maxilla, which develops after sinus surgery or traumatic dental extraction. In very rare cases, it may occur in the mandible . 10 caseIn our case, the cyst was detected in a 37-year-old woman that presented with mild swelling without fistula formation. Review of the literature showed that of the 11 cases that had the cyst in the mandible, 5 were female and 6 were male with an age of onset of 24-72 years . Its rea, and Lefort surgery in men aged 42-72 years. The second man had a history of surgery 56 years ago and the first man underwent surgery 8 years ago. Some symptoms such as painless swelling and numbness of the chin were mentioned as the symptoms of this cyst (In 2014, Li et al reported two cases of ciliated surgery cyst in the mandible after rhinoplasty, genioplastyhis cyst -12. Howehis cyst , it was In our patient, because the ciliated cyst developed in the mandible, the term ectopic ciliated cyst was used. Therefore, long-term follow-up of patients who undergo cosmetic or genioplasty surgery is necessary due to the probability of the formation of such cysts. Enucleation is performed in most patients who have an ectopic ciliated cyst in the mandible. Ragsdale suggested enucleation and peripheral ostectomy, and Li et al suggested abortion .Ectopic ciliated cyst is a rare non-odontogenic cyst in the mandible following genioplasty and orthognathic and Lefort surgery. Therefore, long-term follow-up of both jaws is recommended after cosmetic surgery.


To investigate the effect of birth weight on infant mortality, illness and care seeking in rural Ghana.Using randomized controlled trial data, we compared infants weighing 2.00–2.49, 1.50–1.99 and < 1.50 kg with non-low-birth-weight infants. We generated adjusted mortality hazard ratios (aHR), adjusted illness rate ratios (aRR) and adjusted odds ratios (aOR) for health-facility admissions and absence of care seeking for four time periods: infancy, the neonatal period, early infancy and late infancy – represented by ages of 0–364, 0–27, 28–182 and 183–364 days, respectively.Among 22 906 infants, compared with non-low-birth-weight infants: (i) infants weighing 2.00–2.49, 1.50–1.99 and < 1.50 kg were about two , eight  and 25  times more likely to die in infancy, respectively; (ii) those born weighing < 1.50 kg were about 48  and eight  times more likely to die in the neonatal period and late infancy, respectively; (iii) those born weighing 1.50–1.99 kg  or < 1.50 kg  had higher neonatal illness rates; and (iv) for those born weighing 1.50–1.99 kg, care was less likely to be sought in the neonatal period  and early infancy .For low-birth-weight infants in Ghana, strategies to minimize mortality and improve care seeking are needed. Our primary objective was to determine the extent to which low-birth-weight infants were at increased risk of mortality and illness in the first year of life. Our secondary objectives were: (i) to assess, among sick infants, the association between birth weight and care seeking and health-facility admissions; (ii) to examine how the effects of birth weight on infant illness and mortality varied between the neonatal period, early infancy and late infancy: and (iii) to investigate whether any effect of birth weight on mortality varied by distance to the nearest health facility and/or socioeconomic status.,The Neovita trial was conducted at the Kintampo Health Research Centre in rural Ghana.Using calibrated electronic or spring scales, fieldworkers recorded birth weights to the nearest 0.1 and 0.2 kg, respectively. Over 70% of enrolled infants were weighed within 24 hours of delivery and only five (0.2%) were weighed more than 72 hours after delivery.,During both pregnancy surveillance and at enrolment, fieldworkers asked each pregnant woman the date of her last menstrual period. Various household, infant and maternal characteristics were recorded at enrolment.The primary outcomes were illness and mortality in the first year of life. For each reported episode of illness, we investigated absence of care seeking and admission to a health facility.If an end date but no start date was recorded for an illness, the start date was assumed to have been five days, i.e. the median duration of illness recorded, before the end date. If both start and end dates were unrecorded, the start date was assumed to be the midpoint between the date the illness was reported and the date of the previous follow-up visit. Similarly, if a study infant died within a year of birth but its date of death was not recorded, it was assumed to have died at the midpoint between the first report of its death and the last report of it being alive. Facility admissions occurring within 28 days of a previous admission were reviewed to assess whether they represented responses to a single ongoing illness.,,The primary exposure, birth weight, was divided into non-low-birth-weight, i.e. at least 2.50 kg, and three categories of low-birth-weight: 2.00–2.49, 1.50–1.99 and less than 1.50 kg.n = 16 398) or inconsistently reported (n = 1935). Given this and the known discordance between mothers’ reports of the dates of their last menstrual periods and the corresponding gestational ages assessed by ultrasonography,–For most of the 28 498 pregnancies identified during the study, data on last menstrual period were missing  364 days of age; (ii) the date of death, for infants who died when aged less than 365 days; or (iii) the last date the infant was seen alive if they exited the study when aged less than 365 days.All analyses were conducted using Stata version 13.1 .We generated Kaplan–Meier curves of the probability of survival for low-birth-weight infants compared with non-low-birth-weight infants. We calculated mortality rates for the first year of life. As mortality rate changes rapidly, particularly in the neonatal period, we used multivariable Cox regression to calculate adjusted hazard ratios (aHR) for the association between birth weight and mortality.To allow for repeated illness events, we used random-effects Poisson regression to calculate adjusted rate ratios (aRR) for the association between birth weight and infant illness.Among infants with reported illness, for each illness episode, we used random-effects logistic regression to calculate adjusted odds ratios (aOR) of the association between birth weight and an absence of care seeking or health-facility admission. For each analysis, we assessed whether the effect of birth weight varied between the neonatal, early and late infant periods – represented by ages of 0–27, 28–182 and 183–364 days, respectively – by fitting birth weight as an interaction term with time period. Similarly, for mortality, we assessed whether the effect of birth weight varied by distance to the nearest health facility and/or socioeconomic status.For all analyses, we used likelihood ratio tests and 95% confidence intervals (CI) to assess the statistical evidence for an association between birth weight and each outcome. We also adjusted a priori for infant sex and single/multiple birth, maternal age, education, illness and occupation and household exposure to indoor smoke, distance to nearest health facility, ethnicity, number of children in family, religion and socioeconomic status.Ethics approval for the collection of data included in this study was granted by the Ethics Committees of the World Health Organization, the London School of Hygiene & Tropical Medicine and the Kintampo Health Research Centre.Of the 22 955 infants enrolled in the trial, we included the 22 906 (99.8%) with complete covariate data in our analyses. Of these, almost 16% (3584) were low-birth-weight .Of the included infants, 698 (3.0%) died younger than 365 days. Of these 698 deaths, 277 (39.7%) occurred in the neonatal period, 248 (35.5%) in early infancy and 173 (24.8%) in late infancy. The numbers of deaths per 1000 live births were 30.5 overall and 22.4, 48.6, 160.0 and 402.0 among infants born weighing at least 2.50, 2.00–2.49, 1.50–1.99 and less than 1.50 kg, respectivelyP < 0.0001). After adjusting for all potential confounders, infants born weighing 2.00–2.49, 1.50–1.99 and less than 1.50 kg were about two (aHR: 2.13), eight (aHR: 8.21), and 25 (aHR: 25.38) times more likely to die in their first year of life than non-low-birth-weight infants . Compared with non-low-birth-weight infants, infants born weighing 1.50–1.99 kg (aRR: 1.57) and less than 1.50 kg (aRR: 1.58) had higher illness rates in the neonatal period – although there was little evidence of an association later in infancy . However, in each of these time periods, an absence of care seeking was only observed for infants born weighing 1.50–1.99 kg.We observed evidence of an absence of care seeking for infants born weighing 1.50–1.99 kg  – compared with ill non-low-birth-weight infants . AlthougP = 0.1383).Overall, 4187 admissions were reported for 3485 infants. We found no association between birth weight and admissions over the first year of life  or in th2χ tests  – sought professional medical care – i.e. from a clinic, doctor, hospital, nurse or pharmacy. There was little evidence of differences – between the low-birth-weight and other infants who died – in the duration of illness, in the time to seek care, in the proportions of families who sought care from non-medically trained sources and in the proportions of infants who were admitted to a health facility or who died in a health facility  of the 698 infants who died. Families of the low-birth-weight infants who died were less likely to have sought care than those of non-low-birth-weight infants who died .We found that the families of sick low-birth-weight infants were less likely to have sought care for their infants, even when those infants were suffering from illnesses that led to their deaths. We are aware of only one study that has investigated this topic: an analysis of 840 infants, in rural Malawi, in which preterm and term infants were found to have accessed health care a similar number of times when investigated at 12, 18 and 24 months of age (Several factors may explain why, in a population where birth weight was strongly associated with death, there was little association between birth weight and either care seeking or admissions for illness. First, our data on illness and any associated care seeking and admissions were based on maternal recall. As caregiver recognition of childhood illnesses in low- and middle-income settings is often poor,Compared with general illness, health-facility admission is probably a more notable event that is less likely to be underreported and is a useful marker for severe disease. Although severity is recognized as an important determinant of care seeking,Another possibility is that, during fatal illnesses, sudden illness onset and rapid disease progression are relatively common among low-birth-weight infants – leaving insufficient time to seek care before death. Although our analyses of fatal illnesses indicated that birth weight had no impact on illness duration or the time taken to seek care, the power of these analyses was limited by the small sample size.This study has several strengths. Given the population-based nature of the cohort and the low numbers of individuals excluded from the analyses, our results are likely to be largely representative of the study area’s population. The large sample size provided sufficient power for us to generate estimates of mortality for several categories of low birth weight, including birth weights of less than 1.50 kg. Our study further benefited from low rates of loss to follow-up and from the almost complete data on mortality, including date of death. Any misclassification of deaths by time period should have been negligible.This study has some limitations. As we lacked accurate data on gestational age at birth, we could not generate separate estimates by level of prematurity or for infants that were small for gestational age. We had fairly robust – albeit recall-based – data on whether care was sought for an infant during illness. However, data on several factors that could differ according to birth weight – e.g. time to illness onset, time to care seeking, disease severity and the type of care sought – other than admission to a health facility – were not collected. These factors merit further study. Despite the large sample size, the study was not sufficiently powered to detect moderate differences in illness or admissions by birth weight, especially in the smaller categories of birth weight stratified by time period. The recall of dates of illness was generally poor – e.g. the start dates of more than 40% of reported episodes of illness were recorded as unknown. Although the assumptions we made in estimating the missing dates may have led to some misclassification by time period, the monthly data collection meant that almost 75% of all imputed illness dates were reported within 30 days of a previous visit.In conclusion, strategies to minimize neonatal and infant mortality should target the entire first year of life of low-birth-weight infants. Care for such infants needs to be improved. Our study highlights the need for further studies in Africa to investigate the association between birth weight and infant illness and mortality and any related caregiving and care seeking. Qualitative research on the care of low-birth-weight infants, including the barriers to – and facilitators of – care seeking, is needed.


Five visits were conducted and weight and length were measured at each visit. Mixed model was employed to analyze the growth pattern of Tibetan infants and its comparison to the Han infants. Propensity Scores (PS) technique was adopted to control for the potential confounding factors. The mixed model found that the birth weight/length had a negative impact on the increment of Tibetan infants after birth . The weight increment of Tibetan infants was greater than Han infant with age , after using PS as a covariant. And another mixed model in which PS was used as a matching factor found similar trend. Compared with Chinese Han infants, Tibetan infants were lower weight and shorter length within one year after birth but they had greater increment of weight, suggesting that Tibetan infants might have a significant catch-up growth within the first year of life.Studies on growth pattern of Tibetan infants and the difference from other child groups were limited due to its special living environment and unique customs. In this study, 253 Tibetan infants were followed-up from their birth to 12 The special geographical environment has constructed the unique Tibetan cultural and social environment. Farming and animal husbandry was the main economic activities. The rural population account for 87% of the total population and >95% of people belong to the Tibetan ethnic group6A cohort study design was adopted in this study. The Tibetan infants born in the project of the multi-micronutrients supplementation during pregnancy in Tibet were included as the subjects. The project’s objective was to improve nutritional status of Tibetan pregnant women and birth outcome and it was conducted in Dazi and Qushui county of Lhasa area. The two counties were selected on the basis of available information on socio demographic factors, population size, and prevalence of childhood malnutrition. Both counties shared many similar characteristics with other rural areas of Tibet and did not present any exceptional features. Considering the difficulty of follow-up due to disperse distribution of inhabitation and feasibility of management, some of the newborns were selected to be followed up. In rural Tibet, farmers lived very scattered so it was difficult to follow the newborns living in the more remote area with higher altitude and inconvenient transportation on time. According to the research plan, two townships with better ability of Maternal and child health management were selected from each county. In these townships, the infants born in the project were followed as far as possible. However, the infants selected to be followed presented similar background characteristics compared with the remaining infants including birth weight and length, parental educational level and family economic status, according to statistical analysis.th month between August of 2010 and December of 2012. Five visits during the follow-up were conducted at their birth, 1st month, 3th month, 6th month and 12th month, respectively. Each visit occurred at the required visiting date (±10 days). Anthropometry of the subjects, as well as information on their feeding and immunization practice was collected at each visit. And the demographic and socioeconomic information of their families were ascertained at birth.Total 1418 pregnant women had normal delivery with live infants, of whom 253 were selected for following-up from their birth to 12st month to 12th month after birth, trained staff went to the infants’ home for measuring the weight and length. Infant recumbent length was measured in light clothes and to the nearest millimeter by using the infantometers . Weight was collected by using an electronic infant scale  when the infants were barefooted and bareheaded. Average weight and length from double measurements were used in analysis. The mother’s weight and height was measured only once at the baseline visit to the nearest millimeter by using adults’ boards . Ponderal Index (PI) was used here to measure the soft tissue growth of the infants and expressed as weight (kg)/length(m)3.Anthropometry data on Tibetan infants and their mothers were collected by the trained maternal and child health care (MCH) staff. The infants’ birth weight and length were measured and recorded from the hospitals in which they were born. From 1There were other variables used in the data analysis. The family demographic and socioeconomic information were ascertained at baseline including parental age and education, family size and economic status and so on. A face-to-face interview was conducted for the mothers and their children to collect related data on child health care, child morbidity, feeding information and mother’s prenatal care and so on at each visit. A designed questionnaire was used for data collection. When necessary, the fathers were asked to help the mothers provide family background information. All information was collected by a trained MCH staff. The mother was underweight as Body Mass Index (BMI) <18.5, normal as 18.5 ≤ BMI < 24 and overweight or obesity as BMI ≥ 24. Household wealth index for infants’ family was established from durable consumer goods of households including bicycle, motorcycle, washing machine, livestock and poultry, orchard and characters of house including types of house and toilet and floor materials using Principal component analysis. Four components were retained which cumulative contribution was 73.8%. Wealth index of each family obtained by contribution rate of each component multiplied variables above. Then the family’s living standard was categorized into rich, middle, poor according to their descending value of wealth index at three equal cutoffs.The interviewers  who can communicate freely in both Tibetan native language and Mandarin were trained and questionnaire administration and anthropometric measurements were standardized before the follow-up. During the follow-up, a checking system was applied including checking in the field by interviewer themselves, checking by officials in the counties and then by supervisor in the project office in Lhasa. Personnel from local hospitals and health offices assisted in organization of investigation and explanation of procedures to subjects. Subjects were re-interviewed when transcription or logical questions and missing values were found. Apparatuses for measurement were checked and/or adjusted for accuracy before daily work. All subjects were interviewed face to face to collect anthropometric data.Declaration of Helsinki and was approved by Ethics Committee in Medical Research, Xi’an Jiaotong University (No. 20070712). A written informed consent to participate in the cohort study was obtained from each mother of the infant.The study was conducted in accordance with the s) or median was used for description of continuous variables. The t-test and χ2 test were performed for comparative analysis. The longitude data on weight and length from the infants in Bolivia (3600 m) and Dutch were used as comparisons. Consideration of longitude data and its strong tolerance to missing values, mixed model was adopted to analyze the change of growth from the birth to 12th month. Two mixed models were established. Firstly, a mixed model was used to explore the change of growth by gender among Tibetan children. In this model, the increment of weight/length between each visit and the birth was regarded as dependent variable and independent variables were gender, age, as well as the interaction of gender and age. Birth weight and length as adjustment was also in the model. Subjects were regarded as random variable. The model was indicated as following:Database was established using Epi data 3.1 software  and duplication was adopted for data entry. Percentage was used to describe categorical variables. Mean and standard deviation  technique. Data from 1388 Han infants used in this compared analysis was from a follow-up study conducted in two counties in western China, Bin and Changwu from 2003 to 2008y was the increment of both Tibetan and Han weight/length, β0 was the intercept, β1 was coefficient of the ethnic, β2 was coefficient of the age, β3 was coefficient of the interaction of ethnic and age, PS was the PS value of each infant, τ was the random effects, e was the standard errors. Height/weight at birth, gender, altitude and PS were controlled.where Moreover, PS was used as a matching factor, which meant that Tibetan and Han infants were matched by their PS and a matched subsample was used for the comparison in the growth pattern between them. As a sensitive analysis, a new mixed model was established, which was the same as the model above except no PS included. All the analysis was accomplished by Stata version 12.0 . Significance of all hypothesis testing was determined at P < 0.05 with two-tailed.st month, 207 at 3rd month, 240 at 6th month, and 233 at 12th month. The information of all the infants at birth was collected at the local hospitals. Of the sample, 129 were males and 124 females. The infants’ fathers were aged 27.4 years averagely (19.4~51.5 years old) and had 5.6 average schooling years. The mean age of their mothers was 26.5 years (17.5~43.5 years old) and the mean schooling years were 5.3 years. The mean height and weight of the mothers was 162.12 cm and 53.49 kg, respectively. About 15% of the mothers were underweight in early pregnancy. A Tibetan family had 6 members on average. Differences between both genders in the social-demographic characteristics were not statistically significant (P > 0.05)  .th month. The mean birth weight was 3.02 ± 0.51 kg, 3.07 ± 0.47 kg for male and 2.97 ± 0.55 kg for female. The weight were doubled at half a year after birth, and tripled at 12th month. The average birth length of the infants was 48.33 cm, male 48.55 cm and female 48.12 cm, and the length increased by 15 cm after six months and by more than 23 cm after a year. The greatest increment for both weight and length occurred during the first month after birth and declined with increasing age. So did the PI increment. While not like weight and length, the highest value of PI was at 3rd month then declined with month  regardless of male or female infants, but there was no gender difference found in the increment, and also there was no interaction between age and gender. The weight increment of male infants on average was 1.19 kg (95%CI: 0.95–1.43) at 1st month and 6.28 kg (95%CI: 5.97–6.59) at 12th month while it was 1.09 kg (95%CI: 0.88–1.31) and 6.28 kg (95%CI: 6.01–6.56) for female respectively. The length increment was 4.93 cm and 4.61 cm for male and female at first month and 23.61 cm and 24.23 cm at 12thmonth. Moreover, the birth weight/length had a negative impact on their increment after birth, which meant if an infant was longer or heavier at birth, it would have slower increment in length and weight .The th month after birth between Tibetan and Chinese Han infants, the mixed model was established controlling for potential factors including gender, birth weight/length, altitude and PS. In the model, the interaction of ethnic and age was involved. The results of PS-adjusted model showed that the weight increment of Tibetan was greater than Han infant with age , while the length increment had no such trend , after controlling covariant PS, gender, altitude and birth weight/length for both models , while the length increment had no such trend , which were similar to the results from the PS-adjusted model , but there was no difference in birth height between male and female infants . After controlling for the birth weight and length, the increment of weight and length of Tibetan infants increased rapidly with age. No interaction was found between age and gender, implying that the physical growth between male and female infants was similar in the first year of their life. This result was different from the Chinese research on Han children8The mean birth weight of Tibetan infants was 3.02 kg and grew by 0.54 kg per month on average. Meanwhile, their birth length was 48.33 cm and there were an increase of 1.93 cm per month averagely. The biggest increment of both weight and length occurred at first month after birth. The birth weight of the male was slightly higher than that of the female  compared with their Han counterparts living in plain of China and thereafter the weight and length of Tibetan infants at each visit was lower than that of Han infants. However, after controlling for potential confounders of socio-demographic factors by using PS as a co-variable in the mixed model, the weight increment of Tibetan infant was greater than Han infant with increasing age, but no such trend was found in length increment. We also adopted analysis of matched sample based on PS as a sensitive analysis. Results was similar to those from the mixed model adjusted for PS, which implied that although Tibetan infants had a lower birth weight, their increment was higher than Han infants within the first year after birth. This suggested that Tibetan infants were smaller in size at birth, but they had significant catch-up growth after birth, mainly weight. The increment of length of Tibetan infants was also increased with age, but showed no difference with Han infants. It might result from the shorter follow-up conducted in this study247rd month, similar to the Han infants2 transport suggested that the Tibetans may be better adapted as judged by less fetal growth retardation and may utilize maternal O2 transport mechanisms better than those living for a shorter time in highland30321134Ponderal index (PI), a measure of soft tissue growth, was involved as an indicator in our study. It reflected the mass of the body, both body fat and muscle mass, may have considerable prognostic value for subsequent growth and development of the newborn27Although this follow-up study provided some significant information on the growth of Tibetan infants, there were some limitations considered carefully. Firstly, constrained by local traffics, language barrier and natural environment and funds, the sample size of the follow-up might be smaller to explore the general principle of the physical growth of Tibetan and further large-scale follow-up studies are required. Secondly, we just followed up the Tibetan children one year after birth and we did not know the growth pattern of these infants after 1 year. The effect of adaption to higher altitudes on the physical growth pattern of Tibetan children aged more than 1 year required further studies. Thirdly, in this study, we compared the growth pattern of Tibetan infants living at higher altitudes with that of Chinese Han infants living in the plain in order to explore the ethnic difference between Tibetan and Han infants. There were lots of factors affecting the growth of children, so we used mixed model to control for possible potential factors which were integrated into propensity scores, but there were still other unobserved confounding factors not included in this study. Moreover, this study did not provide gene study which was might be helpful to explain effect of the adaption to high altitudes on physical growth of Tibetan children.st year of the life. The study implied that Tibetan infants had their own physical growth pattern different from Han infants and the adaption to highland environment could play an important role for the ethnic difference. Therefore, for comparison and assessment of physical growth status of Tibetan children, the effect of ethnicity on growth should be taken into account.Although Tibetan infants had the lighter birth weight and shorter birth length, they had higher birth PI, which reflected a better body mass for Tibetan infants. Compared with Chinese Han infants, Tibetan infants were lower weight and shorter length within one year after birth but they had greater increment of weight which suggested that Tibetan infants had a significant catch-up growth with 1How to cite this article: Wang, W. et al. The Growth Pattern of Tibetan Infants at High Altitudes: a Cohort Study in Rural Tibet region. Sci. Rep.6, 34506; doi: 10.1038/srep34506 (2016).


Paraurethral cysts are usually asymptomatic and frequently detected incidentally during routine pelvic examination, however, patients can present with complaints of a palpable cyst or with lower urinary tract symptoms (LUTS) and also dyspareunia. In most cases, diagnosis can be made on physical examination but for more detailed evaluation and to differentiate from malign lesions ultrasonography (US), voiding cystourethrogram (VCUG), computerized tomography (CT), or magnetic resonance imaging (MRI) can also be used. Management of symptomatic paraurethral cyst is surgical excision.In this video our objective is to show the surgical management of female paraurethral cyst with concomitant stress urinary incontinence (SUI).A 37 year-old woman presented with an 8-year history of progressive urinary symptoms, consisting of dysuria, urinary frequency, urgency urinary incontinence, SUI and dyspareunia. Physical examination in the lithotomy position revealed a cystic lesion located in the left anterolateral vaginal wall. Also cough stress test for SUI was positive. Her preoperative ICI-Q, UDI-6, IIQ-7 and SEAPI scores were 16, 8, 9 and 18 respectively. Vaginal US revealed a solitary 2 cm paraurethral cyst, localized in the distal urethra. Pelvic MRI also revealed a benign cystic lesion in the distal urethra. The patient underwent surgical excision of the cyst and anterior colporrhaphy for SUI. At third month visit the patient was very satisfied. The ICI-Q, UDI-6, IIQ-7 and SEAPI scores were 0.Sometimes the LUTS concurring with the parauretral cyst can be dominant. Herein we want to show that extra surgical procedures can be necessary with paraurethral cyst excision for full patient satisfaction. Available at: http://www.intbrazjurol.com.br/video-section/20140582-yonguc_et_al Int Braz J Urol. 2017; 43 (Video #18): 1194-1194


Sclerotherapy and surgery are both effective treatment methods for lymphatic malformations.We report a case of a 12-year-old male patient who presented with a mainly subcutaneous mixed lymphatic malformation located on the right flank . After Video 1Under sonographic guidance, the largest cyst of the lymphatic malformation was cannulated, and an 8 French pigtail catheter was advanced into the cyst over a guidewire. The cyst was filled with 0.9% sodium chloride solution to facilitate the placement of two 3-mm trocars into the cyst. Intralesional endoscopy showed a mixed macro-/microcystic lymphatic malformation with hemorrhage. Dissection of intercystic septa was performed under endoscopic visualization using hook electrocautery and blunt dissection. At the end of the procedure, a single macrocystic cavity had been artificially created. Picibanil (OK-432) was instilled through the formerly inserted pigtail catheter into the now solitary cyst and left in situ for 24 hours.Intralesional endoscopy and intercystic septectomy constitute an interesting novel approach for the diagnosis and treatment of mixed lymphatic malformations. By creating a single, communicating cavity, the efficacy of subsequent sclerotherapy may be increased, and thus the need for multiple treatment sessions may be avoidable.


We provide evidence that substantial variance in white matter microstructure is shared across a range of major tracts in the newborn brain. Based on diffusion MRI scans from 145 neonates [gestational age (GA) at birth range 23+2–41+5 weeks], the microstructural properties of eight major white matter tracts were calculated using probabilistic neighborhood tractography. Principal component analyses (PCAs) were carried out on the correlations between the eight tracts, separately for four tract-averaged water diffusion parameters: fractional anisotropy, and mean, radial and axial diffusivities. For all four parameters, PCAs revealed a single latent variable that explained around half of the variance across all eight tracts, and all tracts showed positive loadings. We considered the impact of early environment on general microstructural properties, by comparing term-born infants with preterm infants at term equivalent age. We found significant associations between GA at birth and the latent measure for each water diffusion measure; this effect was most apparent in projection and commissural fibers. These data show that a latent measure of white matter microstructure is present in very early life, well before myelination is widespread. Early exposure to extra-uterine life is associated with altered general properties of white matter microstructure, which could explain the high prevalence of cognitive impairment experienced by children born preterm.A latent measure of white matter microstructure (The online version of this article (doi:10.1007/s00429-017-1455-6) contains supplementary material, which is available to authorized users. White matter tracts connecting cortical networks are fundamental substrates of higher cognitive function in humans. ‘Disconnection’ of networks, which can be inferred from the microstructural properties of tracts, characterizes a number of diseases and contributes to functional impairment through reduced information transfer efficiency  in one tract is likely to have high FA in all other tracts in the brain). This property allows for the derivation of a general factor, D›, axial (λax) and radial (λrad) diffusivities, and FA, for the major white matter fasciculi during early brain development   is an automatic segmentation technique based on single seed point tractography, that can identify the same fasciculus-of-interest across a group of subjects by modelling how individual tracts compare with a predefined reference tract in terms of length and shape  is present in the newborn; second, preterm birth is associated with global disconnectivity; and third, that g measured in the newborn period is associated with emergent social cognitive function in infancy.We tested the following hypotheses: first, a latent measure of general white matter microstructure (+2–41+5 weeks) were recruited from the Royal Infirmary of Edinburgh between February 2013 and August 2015 to a longitudinal study of the effect of preterm birth on brain structure and long-term outcome. Infants had diffusion MRI (dMRI) at term equivalent age  and 83 took part in eye-tracking assessment 6–12 months later .145 neonates , and healthy controls recruited from postnatal wards with GA 37–42 weeks (n = 36). Exclusion criteria included major congenital malformations, chromosomal abnormalities, congenital infection, overt parenchymal lesions  or post-hemorrhagic ventricular dilatation. Demographic information is shown in Table To study the effect of preterm birth on white matter microstructure the group was divided into those with GA at birth <35 weeks n = 09, and hOf the preterm group: 7% had intra-uterine growth restriction (IUGR) defined as a birth weight under the third centile for gender and gestation and 31% had bronchopulmonary dysplasia defined as need for supplementary oxygen at 36 weeks’ PMA. PMA; postmenstrual age.3, and acquisition time = 7 min 49 s); T2-weighted SPACE ; dMRI using a protocol consisting of 11 T2- and 64 diffusion-weighted (b = 750 s/mm2) single-shot spin-echo echo planar imaging (EPI) volumes acquired with 2 mm isotropic voxels (TE = 106 ms and TR = 7300 ms). Infants were scanned without sedation in natural sleep using the feed-and-wrap technique. Physiological stability was monitored using procedures described by Merchant et al.  and 12-channel phased-array head coil were used to acquire: T1-weighted MPRAGE , tract-averaged values were derived from eight major fasciculi segmented using probabilistic neighbourhood tractography (PNT) optimized for neonatal dMRI data  to extract the brain and eliminate bulk patient motion and eddy current-induced artifacts by registering the diffusion-weighted to the first T2-weighted EPI volume of each subject. Using DTIFIT, MD and FA volumes were generated for each subject. From the underlying white matter connectivity data, eight major white matter fasciculi thought to be involved in cognitive functioning were segmented: genu and splenium of corpus callosum, left and right cingulum cingulate gyrus (CCG), left and right corticospinal tracts (CST), and left and right inferior longitudinal fasciculi (ILF). As described in detail in the study by Anblagan et al.  across the eight tracts, to quantify the proportion of shared variance between them . That is, four separate data matrices (one for each DTI parameter) were separately analysed, each with dimensions n × m where n = 145 (number of subjects) and m = 8 . Thus, each PCA included data from all participants, and all available tracts were included; where tract data were missing , the mean FA, MD, λax and λrad of the group was used to impute values for the missing tract. Next, we examined the effect of preterm birth on differences in these four general water diffusion measures. Initially, we used a dichotomous group design, comparing differences between preterm infants’ and controls’ white matter microstructure (corrected for age at MRI scan) using Welch’s unpaired t tests. We then applied linear regression across the entire group to quantify the dose effect of birth term on each measure of microstructure, including PMA at MRI scan and sex as covariates in the model. To compare tract loadings (correlations between the manifest variable and extracted component score) for each tract between preterm infants and controls, we used Fisher’s test of correlation magnitude differences among independent groups (cocor.indep.groups in the cocor package in R) (Diedenhofen and Musch https://www.r-project.org) version 3.2.2 (Fire Safety).One principal component analysis (PCA) was conducted for each of the four water diffusion parameters , and 36% (λax) of the variance ; gMD ; gλrad ; gλax , Fig. There were significant differences in gFA), β 0.305 (p < 0.001); MD (gMD), β −0.351 (p < 0.001), λrad (gλrad) β −0.363 (p < 0.001); and λax (gλax) β −0.300 (p < 0.001)  at birth and general measures of: FA (01) Fig. . In summZ broadly confirmed this pattern for genu  and left CST , though differences were not significant in the splenium  and right CST . This pattern was also present for λrad and MD, though statistical tests indicated only trend-level or weaker differences for λrad  with the differences in the same direction for MD being smaller and non-significant.In view of variations in newborn network connectivity , and nor were there any significant effects of group within the model  or splenium  of the corpus callosum. Social preference scores for each task  are shown in Supplemental Table 1.There were no significant associations between any general water diffusion parameter and a sensitive measure of emergent social cognition derived from an eye-tracking task battery at 7 months Below is the link to the electronic supplementary material.


Incense burning for rituals or religious purposes is an important tradition in many countries. However, incense smoke contains particulate matter and gas products such as carbon monoxide, sulfur, and nitrogen dioxide, which are potentially harmful to health.We analyzed the relationship between prenatal incense burning and birth weight and head circumference at birth using the Taiwan Birth Cohort Study. We also analyzed whether the associations varied by sex and along the distribution of birth outcomes.We performed ordinary least squares (OLS) and quantile regressions analysis on a sample of 15,773 term births  in Taiwan in 2005. The associations were estimated separately for boys and girls as well as for the population as a whole. We controlled extensively for factors that may be correlated with incense burning and birth weight and head circumference, such as parental religion, demographics, and health characteristics, as well as pregnancy-related variables.Findings from fully adjusted OLS regressions indicated that exposure to incense was associated with lower birth weight in boys  but not girls . Associations with head circumference were negative for boys  and girls . Quantile regression results suggested that the negative associations were larger among the lower quantiles of birth outcomes.OLS regressions showed that prenatal incense burning was associated with lower birth weight for boys and smaller head circumference for boys and girls. The associations were more pronounced among the lower quantiles of birth outcomes. Further research is necessary to confirm whether incense burning has differential effects by sex.Chen LY, Ho C. 2016. Incense burning during pregnancy and birth weight and head circumference among term births: The Taiwan Birth Cohort Study. Environ Health Perspect 124:1487–1492; http://dx.doi.org/10.1289/ehp.1509922 Incense2), nitrogen dioxide (NO2), and volatile organic compounds (VOCs) such as benzene, toluene, and xylenes, which are potentially harmful to health , gas products such as carbon monoxide (CO), sulfur dioxide  is a longitudinal survey that follows 21,248 representatively sampled infants born in 2005 in Taiwan. The study was approved by the Institutional Review Board of Bureau of Health Promotion, the Department of Health and the Directorate-General of Budget, Accounting, and Statistics, Executive Yuan, ROC (No. 94-C3–0940005257), and informed consent was obtained from all participants. The study was conducted through face-to-face interviewer-administered questionnaires .We used data from the first-wave interview, which contained information on pregnancy inputs, infant birth outcomes, parental health conditions before and during pregnancy, parental education, religion, and other demographics. Birth weight and head circumference variables were obtained from birth records, and the remaining variables used in our analysis were self-reported. We limited the sample to singleton (561 observations dropped) and full-term babies , whose parents were both 18–54 years of age (306 observations dropped), and whose parents were both Taiwanese . We dropped observations with missing values for birth weight (26 observations), incense burning (3 observations), maternal smoking during pregnancy (5 observations), maternal health indicators (58 observations), number of previous births (65 observations), parental weight and height (242 observations), parental chronic illness (108 observations), marital status (2 observations), and parental religion (8 observations). This left us with a sample of 15,773 infants among whom 8,216 were boys and 7,557 were girls.We focused on two measures of infant health: birth weight (measured in grams) and head circumference at birth (measured in millimeters). We observed whether the household had a habit of burning incense from the question “Does your household have the habit of burning incense for religious purposes?” We constructed a dummy variable with a value of 1 if the household had the habit of burning incense and 0 otherwise. For the sample of all babies and the subsamples stratified by sex, we plotted the kernel densities  for birth weight and head circumference across households that burned incense and households that did not. We also observed whether mothers smoked during pregnancy from the questions “Within the first three months of pregnancy (first trimester), did you smoke?” and “After the fourth month of pregnancy (second trimester), did you smoke?” We constructed a dummy variable taking a value of 1 if she smoked during pregnancy and 0 otherwise.p-value of the interaction term, from two-tailed t-tests, was obtained as regression output .To estimate the associations between incense burning, maternal smoking and birth outcomes, we ran ordinary least squares (OLS) regressions  as well as quantile regressions . The estimated coefficients of incense burning and maternal smoking from OLS were reported as forest plots , and the estimated coefficients of incense burning from the quantile regressions were reported as quantile plots throughout the distribution of birth outcomes . The OLS and quantile regressions were run separately for the outcomes of birth weight and head circumference, and separately on the full population of full-term births, the sample of boys, and the sample of girls. We also included an interaction term between incense burning during pregnancy and being a male baby in OLS regressions separately for each of the aforementioned outcomes on the full population of full term births. The To ensure that our estimates of interest are robust, we modeled each association using three different covariate specifications: model 1, adjusted for demographic variables only; model 2, additionally adjusted for parental health characteristics; and model 3, a fully adjusted model that also included pregnancy-related variables. Demographic characteristics included marital status , mother’s and father’s ages at the time of birth , mother’s and father’s education , mother’s and father’s Chinese religion , and residence area . Parental health variables included mother’s prepregnancy body mass index (BMI) and father’s BMI , maternal hepatitis B status , and any chronic illness in the mother or father . Pregnancy-related variables included any maternal hospitalization during the pregnancy (yes/no), any pregnancy illness , gestational diabetes (yes/no), tocolysis (yes/no), any injury to the abdomen during pregnancy (yes/no), first pregnancy (yes/no), previous abortion , previous miscarriage (yes/no), any children < 6 years of age living in household (yes/no), and number of previous births . The selected covariates were controlled for as they may be correlated with birth outcomes and incense burning.We used an alpha of 5% as the significance level when reporting whether our estimates of interest are statistically significant.Birth weight was observed for 15,773 babies  in the sample, and head circumference at birth was observed for 14,488 babies . The average birth weight for the population of full-term births, for boys, and for girls was 3,166 g, 3,217 g, and 3,111 g respectively. The average head circumference for the population of full-term births, for boys, and for girls was 334 mm, 336 mm, and 332 mm respectively.Fifty-nine percent of households had the habit of burning incense and 3.6% of the mothers in the sample smoked during pregnancy. From kernel density graphs , there was a slightly higher proportion of lower birth weight babies in households that burned incense compared with those that did not burn incense. The evidence was less conclusive for head circumference.The average age of mothers and fathers was 29 and 32 years, respectively. Twenty-three percent of mothers and 28% of fathers were college educated. Approximately 77% of parents followed a Chinese religion such as Chinese folk religion, Buddhism, Taoism, and I-Kuan Tao. Thirty-six percent of mothers were experiencing a first pregnancy, and, on average, mothers had delivered 1.64 live births. Additional information on all variables used in our analysis are reported in The OLS regression results for the outcomes of birth weight and head circumference at birth are reported in p > 0.05) between incense burning and birth weight in the overall sample based on the fully adjusted model  . There was also a negative association between incense burning and average head circumference at birth. Incense burning was associated significantly with smaller head circumference in the full sample of babies  and in boys . The associations were negative but nonsignificant in girls . The interaction p-value from the head circumference regression was larger than that from the birth weight regression, suggesting that sex differences need to be confirmed through future research.There was a negative but nonsignificant association , were within the range of –50 g to –600 g found in the literature on maternal smoking (yes/no) and birth weight . On the 2, NO2, and VOCs, are also present in cigarette smoke  also allowed us to conduct the analysis separately by sex, which seemed to suggest some differences between girls and boys for the outcome of birth weight; however, further research is necessary to confirm sex differences. We also performed mean and quantile regressions, which allowed us to analyze whether the associations differed in means and distributions across birth outcomes.Whereas our rich data set enabled us to estimate the associations between incense burning practice and birth outcomes, we note the following limitations of our study: First, there may be other determinants of birth outcomes that we could not control for due to data limitations. For instance, season of birth is known to affect birth outcomes , so contThere is a growing interest on the relationship between ambient air and health outcomes. In this study, we addressed the issues of incense smoke and infant birth outcomes. Using a nationwide data set from the Taiwan Birth Cohort Study, we estimated the empirical relationship between incense burning during pregnancy, birth weight, and head circumference at birth. Our findings from mean and quantile regressions suggest that the associations between incense burning and infant birth outcomes were generally negative. Moreover, the associations with lower quantiles of the distributions of birth outcomes appeared to be more pronounced than those with the upper quantiles. Our findings need to be confirmed in other populations, including differences between boys and girls.(336 KB) PDFClick here for additional data file.


Despite evidence of between‐arm differences in some fetal biometry, z‐scores at birth were not greater in the intervention arms than the FeFol arm . In regression analyses, intervention associations with birth weight and head circumference were modified by maternal weight gain between booking and 30 weeks gestation  in the highest quartile of weight gain but –0.099 z‐scores  in the lowest). In conclusion, we found no strong evidence that a prenatal LNS intervention was associated with better fetal growth in the whole sample.Prenatal supplementation with protein‐energy (PE) and/or multiple‐micronutrients (MMNs) may improve fetal growth, but trials of lipid‐based nutritional supplements (LNSs) have reported inconsistent results. We conducted a post‐hoc analysis of non‐primary outcomes in a trial in Gambia, with the aim to test the associations of LNS with fetal growth and explore how efficacy varies depending on nutritional status. The sample comprised 620 pregnant women in an individually randomized, partially blinded trial with four arms: (a) iron and folic acid (FeFol) tablet , (b) MMN tablet, (c) PE LNS, and (d) PE + MMN LNS. Analysis of variance examined unadjusted differences in fetal biometry  Given tRecently developed lipid‐based nutritional supplements (LNSs), which are affordable, safe, can be produced locally, and have a long shelf‐life, have been shown to be a very effective option for the community‐based treatment of severe malnutrition if a high dose is given  based on the offspring being live births with complete birth weight and gestational age data. Defining characteristics were not different between this sample and those ENID participants who did not meet the inclusion criteria (N = 875–620 = 255). For example, where data were available, the comparison was 162.0 versus 61.7 cm (p = .442) for maternal height at booking, 29.6 versus 29.9 years for maternal age at booking (p = .642), and 37.9 versus 38.7 for the percentage of births occurring in the raining season (p = .861).The sample comprised 620 mothers and their singleton offspring  enrolled in the early nutrition and immune development trial  in rural Gambia. As shown in Supplementary Figure 2.2The ENID trial has been described in detail elsewhere  were distributed on a weekly basis to participating women. Women were supplied with 14 tablets per week in individual bottles and advised to take two tablets per day, preferably with food. LNS were supplied in jars, with a single (daily) dose per jar. Women were encouraged to consume the whole jar each day, with the option of eating it straight from the jar (spoons were supplied) or eating it with food. Given the common practice of sharing from a family bowl, women were encouraged to take a separate portion of the food from the family bowl and mix the LNS into that. The first women started to receive supplementation in January 2010, and the final infant was born in February 2014. Compliance was assessed through the collection of all unused supplements at the end of each week. For tablets, a count on remaining tablets was performed, and for LNS products, a score based on the amount of supplement left remaining in the jar was made . A compliance percentage was computed for each woman by dividing the number of LNS pots or tablets the woman consumed by the number she was offered, and multiplying by 100.The nutritional composition of the intervention products is detailed in the study protocol published by Moore et al. . BrieflyFrom 6 to 12 months of age, infants were further randomized to a LNS supplement, with or without additional MMN, but this second randomization stage does not need to be considered for the purposes of the present paper as our focus is on fetal growth.The trial was approved by the joint Gambia Government/MRC Unit, The Gambia Ethics Committee (Project number SCC1126v2). Written informed consent was obtained from all women prior to enrolment into the trial. The trial observed Good Clinical Practice Standards and the current version of the Helsinki Declaration. The trial was registered as ISRCTN49285450.This paper uses data from the clinic enrolment or “booking” visit, subsequent clinic visits at 20 and 30 weeks of gestation, and a home visit performed within 72 hr of birth.At the three prenatal clinic visits, fetal biometry was assessed via ultrasound using a Siemens ACUSON Antares Ultrasound Imaging System  Inc; California, USA with a CH6–2 (5.71 MHz) transducer). Using the built‐in equations, the purpose of the fetal biometry measurements taken at booking was to estimate gestational age. This estimation was based on crown‐rump length (CRL) if gestational age was <12 weeks, or bi‐parietal diameter (BPD) if CRL was too large to be accurately measured or gestational age was ≥12 weeks. If CRL was not yet measurable, gestational age was estimated according to the size of the gestational sac, and the woman's booking visit was rescheduled for 12 weeks gestation. Subsequently, at the 20 and 30 week visits measurements of femur length (FL), head circumference (HC), occipital‐frontal diameter (OFD), and abdominal circumference (AC), as well at BPD were performed using standard methods  with the infant in minimal clothing and to the nearest 10 g. Length was measured on a portable infant rollameter  to the nearest 0.1 cm. HC was measured using standard circumference tapes . All measures were made using standard protocols, and equipment was regularly validated.Maternal weight and height were measured at the booking and 20 and 30 week visits using standard techniques and equipment . Maternal date of birth and thus age at booking were ascertained from the West Kiang Demographic Surveillance System ; neonatal anthropometry models additionally included age at assessment and parity. Given our sample size and the hypothesised association modification, logistic models for binary outcomes  were not performed because of the reduced power this approach provides.General linear regression models were used to test whether or not intervention associations  with fetal biometry  and neonatal anthropometry  Z‐scores were modified by the following a priori specified variables: season of assessment, maternal height and BMI at booking, and maternal weight gain from booking. These variables were chosen based on evidence from previous publications that fortified LNS may impact on fetal growth and development most among women who are more nutritionally vulnerable , analyses were rerun using neonatal anthropometry from a week one visit instead of the birth visit.To investigate if compliance or length of time on supplement were affecting the efficacy of the intervention, sensitivity analyses were performed removing individuals in the bottom quartile of compliance (standardized within each arm) or length of time on supplement. Further, given that the majority of exclusions were due to missing birth weight .3Baseline characteristics of the study sample, stratified according to intervention arm, are shown in Table z‐scores: FeFol −0.29, MMN −0.28, PE +0.13, PE + MMN −0.22; group‐wise p = .012). A similar pattern was observed at 30 weeks of gestation, but for different biometry measures . Despite this evidence of between‐arm differences in some fetal biometry measures, likely due to higher values in the PE group, neonatal anthropometry z‐scores were not greater in the intervention arms than the FeFol arm .The key findings are summarized in Table p = .016), although were no between‐arm differences in rates of preterm, low birth weight, and SGA  but not among those who were in the lowest quartile . Similar evidence of association modification by maternal weight gain was observed for HC and WFL, but not length z‐scores.Table Nearly identical results to those presented in this paper were obtained (a) in analyses restricted to individuals in the top three quartiles of compliance (standardized within each arm) or length of time on supplement and (b) in analyses using neonatal anthropometry from a week one visit instead of the birth visit (data not presented).4z‐scores at birth were not greater in the intervention arms than the FeFol arm. Our key finding, therefore, is that prenatal LNS intervention was not associated with better fetal growth in the whole sample. Pronounced seasonality in The Gambia affects many aspects of diet, health, and behavior  persisting to 20 and 30 weeks of gestation than differential compliance between the two LNS arms . The fetal biometry results, therefore, have to be interpreted with caution, particularly given that we did not observe any between‐arm differences in weight, length, or HC at birth in the whole sample.The main strengths of the present study are that (a) supplementation started early in gestation (~13 weeks), which is important knowing that even pre‐conceptional nutritional status may impact on birth weight  with birth weight and HC among women who demonstrated the greatest gestational weight gain.This trial is supported by the UK Medical Research Council (MRC) (MC‐A760‐5QX00) and the UK Department for International Development (DFID) under the MRC/DFID Concordat agreement. WJ and SEM are funded by the UK MRC programme MC_UP_1005/1.The authors declare that they have no conflicts of interest.The authors' responsibilities were as follows: SEM, MKD, and AMP conceived and designed the ENID Trial; MKD, FS, PN, and SEM conducted the research; WJ and SEM conceived this post‐hoc analysis; WJ performed the statistical analyses and drafted the manuscript. All authors critically revised and approved the manuscript.


Measurements and models of electric fields in the Published 07, February 2017Since publication of the paper, Alex Opitz alerted us of a problem with the calibration of the voltage recordings. When stimulating with sinusoidal currents the Neuroconn stimulator that was used in our experiments specifies stimulation intensity as peak-to-peak amplitude, which we now confirmed by measuring current through a 10kOhm load. We had incorrectly interpreted this as zero-to-peak amplitude. In the first step of analysis we divided all voltage measures by this nominal current value to obtain measures calibrated to 1 mA of stimulation. Due to this misinterpretation, all values we reported in the paper for measured voltages and field magnitudes have to be multiplied by a factor of 2.in-vivo measurements. Our results now match recent reports from monkey and human of approximately 0.8 V/m maximum field amplitudes in the brain when stimulating transcranially with 2 mA current . Subsequent recordings were performed at 1 Hz and with stimulation current of 0.25 mA to 1 mA, limited by the dynamic range of the clinical amplifier and patient sensation.1. Regardless, the variability observed here sets the lower limit in precision one can expect to 2. For this subject (P03) the measured voltages are tightly correlated across locations with the predicted electric potential values. The same is true for all subjects resulting in high Pearson correlation coefficients, V/m and V/m, respectively).3. As expected, the correlation of predicted and measured electric fields is lower than for the raw potentials . For an intensity of 2 mA typically used in clinical trials this would result in fields of 0.56 V/m. For other stimulation configurations on Subject P014 with electrodes away from the mid-line , maximum intensity on cortex for 1 mA stimulation was 0.25 V/m and 0.10 V/m at the fronto-lateral and lateral occipital locations respectively. We also model other electrode configurations commonly used clinical trials, e.g., M1–SO . For the three configurations tested here, maximum field intensities in cortical locations was 0.38 V/m ± 0.09 V/m, again for 1 mA. To provide a sense of what intensities are reached in more extended areas  Figure 6A also reports the 95th percentile (0.14 V/m ± 0.02 V/m)…4. Maximal field intensities for the mid-line configuration  are 0.05 V/m. … Assuming that prediction accuracy scales linearly with predicted electric field intensity, this suggests that prediction error for maximal field intensities is in the order of 0.10 V/m.5. For a given predicted electric field intensity, measurements vary by approximately V/m for these 10 subjects with the mid-line montage).6. The maximal values at a given depth summarized for all subjects and montages indicate that deep brain areas may experience electric fields that are comparable in magnitude to the cortical surface  field magnitude were appear to give similar estimate of the electric field magnitudes .7. First, conductivity values reported in the literature used in existing models close to the predicted values, as indicated by the slope appear to correctly estimate the amount of current that crosses the skull and the amount shunted through the scalp.8. … measured electric field magnitudes in subject P03 were 0.03 ± 0.01 S/m for bone, 0.29 ± 0.10 S/m for skin, 0.38 ± 0.05 S/m for white matter and 0.82 S/m for gray matter . These differ from the literature values , but are largely in the same proportions.9. The median of the optimal conductivities are but the magnitude is not significantly different  using a standard head that captures the average anatomy of the lower head . This extended FOV gives significantly better predictions as compared to the original FOV, in terms of the electric field magnitudes  under 2 mA stimulation. For some electrode montages, current is shunted into deeper areas through highly conductive CSF with maximal values reaching 0.21 V/m. The present data is broadly in agreement with previous modeling estimates with maximal intensities around 1 V/m . These results are in remarkably close agreement if one considers the differing placement of the stimulation electrodes.4. A recent study has reported the electric field produced by TES as measured by stereotactic EEG electrodes in two epilepsy surgical patients . While t5. After calibrating the models the overall magnitudes of electric field were predicted with a single set of parameters across all subjects without a bias .6. The overall correlation of predicted and measured fields is 0.11 mV, e.g., Figure 3C), and note that the variations we did observe were likely due to subject movement.7. We found relatively little variation in the voltage measurements , with a sinusoidal waveform, at variable frequencies and intensities between http://dx.doi.org/10.6080/K0XW4GQ1.In closing we note that all new analyses and figures exclude duplicate measurement points following the considerations discussed in an earlier commentary. We also adjusted electrode location in a few subjects based on more detailed description from AAL who placed electrodes underneath the bandage. All corrections are also reflected in the data repository


The aim of this study was to establish the reference range for SpO2 in healthy term and preterm neonates within 24 hours in Nairobi, Kenya, located at 1800 m. A random sample of clinically well infants had SpO2 measured once in the first 24 hours. A total of 555 infants were enrolled. The 5th–95th percentile range for preductal and postductal SpO2 was 89%–97% for the term and normal birthweight groups, and 90%–98% for the preterm and low birthweight (LBW) groups. This may suggest that 89% and 97% are reasonable SpO2 bounds for well term, preterm and LBW infants within 24 hours at an altitude of 1800 m.There are minimal data to define normal oxygen saturation (SpO Hypoxaemia occurs in a substantial portion of hospitalised neonates, and is significantly associated with mortality. Pulse oximetry is a non-invasive method of measuring the oxygen saturation of haemoglobin, and has become a critical tool in determining need for oxygen in sick newborns. Routine use may also aid identification of infants with clinically unrecognised respiratory abnormalities in LIC, where discharge often occurs within the first 24 hours of life.2) values. It is important to define normal SpO2 levels for neonates of different gestational ages (GA) at different time points after birth. SpO2 values prior to 24 hours are lower and more variable than those seen after 24 hours. Similarly, SpO2 values may be lower at higher altitudes, such as in Kenya where 20% of the population resides above 1500 m. Thilo et al2 was 92%–93% at 24–48 hours among well term neonates at 1610 m. Ravert et al2 of 95%–97% at ∼1371 m and 94%–95% at ∼2073 m among well term newborns during the first 72 hours. Three studies evaluated SpO2 in healthy preterm infants. Ng et al2 of 97% ; Harigopal et al2 of 95%  and Richard et al2 of 99% . All of these studies were conducted at sea level.2 in healthy term and preterm neonates at 1800 m.Use of pulse oximetry relies on knowledge of normal oxygen saturation SpO values. This study was conducted at Pumwani Hospital in Nairobi, Kenya, located at 1800 m. The hospital provides care to ∼22 000 women and their infants each year. A 120-bed nursery provides care for all infants requiring medical attention. Inclusion criteria included being born alive at Pumwani Hospital during the study period and appearing well as defined by (1) normal vital signs , (2) absence of respiratory distress and (3) absence of other signs of illness . Infants were excluded if they required admission for any reason other than observation or if they were transferred to another facility within 4 hours for anomalies or severe medical problems.We aimed to enrol 800 well neonates, stratified to include 400 term (≥37 weeks) and 400 preterm (<37 weeks) neonates. GA was based on last menstrual period (LMP), and Ballard examination was conducted when LMP was unknown or incongruent with appearance. We screened all preterm neonates and a random proportion of term neonates.2 was stable over that period.Enrolled infants underwent testing once between 1 and 24 hours with the Lifebox oximeter, which was developed for low-resource settings by WHO and World Federation of Societies for Anaesthesiologists . The measurement was recorded when there was a good waveform for ≥15 s and SpO2 measurements, stratifying by GA (term or preterm), birth weight , ≥2.5 kg or low (LBW), <2.5 kg) and postnatal age . Statistical analyses were conducted using Stata V.13 .We determined mean, SD, median, IQR and 5th–95th percentile range of preductal and postductal SpOWritten informed consent was obtained from parents/guardians. Ethical approval was received from the University of California, San Francisco and the Kenya Medical Research Institute-Wellcome Trust Research Programme.A total of 555 infants were enrolled between January and December 2015. The mean GA was 38 weeks (SD 2.4), mean birth weight was 2.94 kg (SD 0.53), 50% were male and 93% were delivered vaginally. Among preterm infants, median GA was 35 weeks (range 27–36).2 was 93%–94% across all GA and birthweight groups. Mean postductal SpO2 was 93%–94% for all GA groups, 93% for birth weights ≥2.5 kg and 95% for <2.5 kg. Median preductal and postductal SpO2 values were similar (2 was between 89% and 97% for the term and NBW groups, and between 90% and 98% for the preterm and LBW groups.Mean preductal SpO similar . The 5thWhen stratified by postnatal age, results were similar .2 by hours are shown in Scatter plots of preductal and postductal SpO2 was 93%–95% in all groups within 24 hours. This is similar to findings among well term infants in studies conducted by Thilo et alet al.2 for preterm infants were only slightly lower than those found by Ng et al,et al2 were 89%–97% for the term and NBW groups and 90%–98% for the preterm and LBW groups. Thilo et al reported a 95% CI of 89% to 97%,Data are sparse to inform our understanding of oxygen saturation norms for newborns within the first 24 hours and for well preterm infants. We found that mean preductal and postductal SpO2 reference ranges in preterm infants born at a variety of altitudes is needed. If methods and timing for collecting measurements were sufficiently similar, these data could make a considerable contribution to a pooled sample for meta-analysis.This study has limitations. Our findings for preterm infants are based on a small sample of 135 infants. It was difficult to recruit well preterm infants as many such infants were admitted for medical therapy, making them ineligible. Additional research about SpO2 was stable over that period.In this study, GA was based on LMP with Ballard examination conducted when LMP was unknown or incongruent with appearance. LMP is subject to bias and Ballard may differ from ultrasound by 1 to 2 weeks. However, ultrasound is unavailable in many LIC facilities due to cost and need for skilled sonographers. The Lifebox oximeter is not motion-resistant and has not been validated in neonates. To ensure we obtained the most accurate measurement possible, we recorded measurements only when there was a good waveform for ≥15 s and SpO2 bounds for well term, preterm and LBW infants within 24 hours at 1800 m, although we lack outcome data that would confirm infants in this study remained well. This study provides important data about SpO2 within the first 24 hours, which are useful as discharge postdelivery is rapid and typically without skilled assessment by a clinician or nurse in many LIC settings.Reports suggest that pulse oximetry use can decrease mortality in children with unrecognised hypoxaemia. In newborns, it is increasingly being used for predischarge screening to diagnose occult respiratory and cardiac disease. This study suggests that 89% and 97% may be reasonable SpO


Optimal birth spacing has health advantages for both mother and child. In developing countries, shorter birth intervals are common and associated with social, cultural, and economic factors, as well as a lack of family planning. This study investigated the first birth interval after marriage and preceding interbirth interval in Nepal.A community-based prospective cohort study was conducted in the Kaski district of Nepal. Information on birth spacing, demographic, and obstetric characteristics was obtained from 701 pregnant women using a structured questionnaire. Logistic regression analyses were performed to ascertain factors associated with short birth spacing.About 39% of primiparous women gave their first child birth within 1 year of marriage and 23% of multiparous women had short preceding interbirth intervals (<24 months). The average birth spacing among the multiparous group was 44.9 (SD 21.8) months. Overall, short birth spacing appeared to be inversely associated with advancing maternal age. For the multiparous group, Janajati and lower caste women, and those whose newborn was female, were more likely to have short birth spacing.The preceding interbirth interval was relatively long in the Kaski district of Nepal and tended to be associated with maternal age, caste, and sex of newborn infant. Optimal birth spacing programs should target Janajati and lower caste women, along with promotion of gender equality in society. Birth spacing is an essential component of family planning and fertility control . OptimalIn developing countries, short birth intervals are prevalent and mostly unintended . Nepal iBesides societal and cultural norms, various factors also contribute to high fertility and short birth spacing at individual, household, and community levels. These include illiteracy, early marriage, lack of family planning methods, and poverty . KnowledThis community-based study was undertaken in Kaski district, located in the relatively developed central hills region of Nepal. The district has a central valley popularly known as Pokhara city. The rural areas stretch out of the valley in hilly terraces. Kaski has 75% female literacy, approximately 134,100 married women with 13,800 expected pregnancies annually .The results were derived from a large prospective cohort study of delivery service utilization and pregnancy outcomes between December 2011 and November 2012. Details concerning the study design and sampling had been described elsewhere , 14. BriThe questionnaire used was adapted from the validated Nepal Demographic and Health Survey. Its validity and reliability had been tested on 50 pregnant women for cultural appropriateness and understanding. Fifteen female data enumerators recruited the pregnant women in their local community and conducted face-to-face interviews at the home of participants to collect information on socio-demographic characteristics, obstetric, and family planning knowledge. Maternal age at recruitment, age at marriage, and the age of their youngest child (for multiparous women) were also recorded.The study protocol was approved by the Human Research Ethics Committee of Curtin University , Ethical Review Board of the Nepal Health Research Council , and the District Public Health Office of Kaski. An information sheet was distributed and read to each participant before obtaining her signed or thumb-print consent. Confidentiality of the information provided was maintained throughout the study.The outcome variable of interest was birth spacing, being the first birth interval after marriage for primiparous women, or the preceding interbirth interval for multiparous women. The former interval was calculated by subtracting the marriage age from the current age, before categorizing as either short (≤12 months) or normal (>12 months). The latter interval was computed using the age of the youngest child and the current maternal age, and then classified as either short (≤24 months) or normal (>24 months). The recommended interval between successive births is at least 24 months, since shorter intervals are known to have negative maternal and perinatal outcomes .t, and chi-square tests. Logistic regression analyses were then performed to ascertain factors associated with the prevalence of short birth spacing. All statistical analyses were undertaken in the STATA package Version 13.Comparisons of characteristics were made between short and normal birth spacing groups using descriptive statistics, n = 363, 51.8%) of the 701 participants were primiparous. Tables n = 338, 48.2%), they were 26.1 (SD 4.0) years and 27 (SD 5.5) weeks, respectively. The average age at marriage was about 19.5 years for all mothers.About half  and knew about family planning methods (69%). Among the multiparous group, information on birth interval was missing for 10 cases. Approximately half of the group was from upper caste and urban area. Most of them had only one previous child (60%) and knew about family planning methods (82%). It is interesting to note that 60% of the newborns in both groups were male.For primiparous women, the mean first birth interval was 19.7 (SD 16.5) months, and 141 (38.8%) of them gave their first child birth within 1 year of marriage. The mean preceding interbirth interval for multiparous women was 44.9 SD 21.8) months with median 42 months. Such birth spacing ranged from 6 to 98 months, with 23.2% being <24 months. The short and normal birth spacing subgroups for both primiparous and multiparous women appeared to be different in terms of caste, knowledge of family planning methods, maternal age, and paternal age  was greater than the average of Nepal (36.2 months)  and thosCaste was another factor affecting the preceding interbirth interval. Lower caste women are typically characterized by less education and wealth . MoreoveNepalese women who gave birth to a female baby appeared to have a shorter birth interval, similar to Ethiopian women , while tFor primiparous women, their first birth interval after marriage was significantly associated with age only. This finding supports that variables affecting conception are parity dependent . Older wSeveral limitations should be considered. In this study, birth spacing for multiparous women was defined in terms of preceding interbirth interval rather than interpregnancy interval, the latter being a sensitive issue and difficult to obtain accurate information since it must account for incidences of abortion. Furthermore, successive birth intervals were calculated based on recall of the age of their youngest child, which might incur recall and reporting biases. Another limitation concerns the lack of information on other confounding factors, such as the use of family planning methods, breastfeeding status, maternal physiological status, gender, and survival status of the previous child.Almost 40% of primiparous pregnant women gave their first child birth within 1 year of marriage, while the birth spacing among multiparous women was relatively longer with a median of 42 months. Most early births (≤12 months of marriage) tend to occur in older, upper caste women who have knowledge of family planning methods. The preceding interbirth interval was associated with maternal age, caste, and sex of the newborn infant. Optimal birth spacing programs should target lower caste and Janajati mothers, along with promotion of gender equality in society.RK conceived and managed the project and data collection, performed statistical analysis, and drafted the manuscript. AL contributed to data analysis and revision of the manuscript. Both authors read and approved the final version for publication.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. The reviewer JC and handling editor declared their shared affiliation, and the handling editor states that the process nevertheless met the standards of a fair and objective review.


The in situ prediction is enabled by coupling MREIT with a corresponding Peleg-Fermi mathematical model to obtain more informative monitoring of IRE tissue ablation by providing cell death probability in the IRE-treated tumors. This technique can potentially be used in electroporation-based clinical applications, such as IRE tissue ablation and electrochemotherapy, to improve and assure the desired treatment outcome.Irreversible electroporation (IRE) is gaining importance in routine clinical practice for nonthermal ablation of solid tumors. For its success, it is extremely important that the coverage and exposure time of the treated tumor to the electric field is within the specified range. Measurement of electric field distribution during the electroporation treatment can be achieved using magnetic resonance electrical impedance tomography (MREIT). Here, we show improved MREIT-enabled electroporation monitoring of IRE-treated tumors by predicting IRE-ablated tumor areas during IRE of mouse tumors  As most theoretical studies have predicted, the increase in cell membrane permeability is a consequence of structural changes that occur in the membrane in the form of hydrophilic pores3. When the cell is exposed to an adequate electric field, transient structural changes can be attained5. After a certain period of time, the membrane reseals and the cell survives. This is termed reversible electroporation as the cell preserves its viability. Medical applications of reversible electroporation, such as gene electrotransfer10 and electrochemotherapy16, exploit these structural changes in the membrane to increase the transmembrane transport of foreign DNA and chemotherapeutic drugs, respectively. In contrast, irreversible electroporation (IRE) induces cell death as the applied electric field is too strong for cells to recover, leading to excessive damage to cells and membranes. In recent years, nonthermal IRE for the ablation of solid tumors20 has emerged as a new medical technique. IRE has also been suggested for the treatment of atrial fibrillation as a nonthermal ablation method with minimal to no local heating effects23.Electroporation is a phenomenon that increases cell membrane permeability due to externally applied pulsed electric fields25. Therefore, the treatment efficacy of electroporation-based treatments, including IRE tissue ablation, can be enhanced by treatment planning, i.e. predicting electric field distribution through numerical modeling30. The outcome of IRE tissue ablation is usually evaluated by post-IRE medical imaging using either magnetic resonance (MR) imaging31, computed tomography32 or ultrasonography33. Even though the latter has been reported to be able to monitor IRE within minutes after intervention34, the main disadvantage of all the above-mentioned methods is the inability to monitor electroporation during pulse delivery. Since the efficiency of electroporation-based treatments is strongly correlated with electric field distribution36, a method capable of determining electric field distribution during the pulse delivery could potentially enable monitoring of electroporation-based treatments. This can be achieved by means of magnetic resonance electrical impedance tomography (MREIT), which enables measurement of electric field distribution during the application of electric pulses using MR imaging and mathematical algorithms38. This method was recently demonstrated in mouse tumors in vivo to predict reversibly electroporated areas through measurement of the electric field distribution during the application of electric pulses39. In that study, however, the authors were only able to show a correlation between the electric field that led to reversible electroporation of tumor cells and the entrapment of contrast agent due to the electroporation. This was done by comparing tumor fractions, i.e., the size of the treated area divided by the size of the whole tumor, and not by direct comparison of either the size or shape of the treated areas. A direct comparison was not possible for technical reasons, such as the considerable time difference (24 h) between the treatment and the assessment of the contrast agent entrapment. Another limitation of that study was that the outcome of the treatment was predicted only based on the applied electric field distribution, i.e., by the amplitude of the electric field. In reality, it is not just the electric field that defines the outcome of electroporation treatments but also the duration of exposure to the field, i.e., the duration of a single electric pulse and the number of the applied pulses41. For this reason, mathematical models have been developed to describe the effects of electroporation on the treated cell suspensions and tissues45. Our present study further improves MREIT-enabled electroporation monitoring of IRE-treated tumors by predicting IRE-ablated tumor areas during electroporation in vivo. This in situ prediction is enabled by coupling MREIT with a corresponding mathematical model to determine cell death probability in IRE-treated tumors.A cell can be exposed to an electric field by the application of electric pulses. Parameters of electric pulses, such as amplitude, duration and the number of pulses, play a major role in determining the outcome of all electroporation-based treatments, including IRE tissue ablationvia two needle electrodes inserted into the tumor. First, a mouse with a tumor was placed in the MRI scanner and then scanned by the current density imaging (CDI) method during application of electric pulses in order to acquire a map of the magnetic field change induced by electric currents. The map was then processed according to Ampere’s law to obtain the corresponding current distribution map, which was then used to calculate the conductivity maps of the tumor using the MREIT algorithm. Finally, when current distribution and conductivity maps were determined, the electric field distribution in the tumor was calculated from the maps using Ohm’s law. Tumor tissue damage induced by IRE was then predicted by the Peleg-Fermi cell-death model using the measured electric field in the tumor and the number of applied electric pulses. After 72 hours, mice were sacrificed, and tumors were excised and fixed in zinc fixative and embedded in paraffin. Sections from each tumor were cut and then stained with hematoxylin and eosin (H&E) in order to determine necrotic areas, i.e. ablated areas due to IRE. The so-determined ablated areas of each tumor were then statistically compared with the corresponding predicted areas for the tissue damage given by the Peleg-Fermi cell-death model.IRE ablation of a murine tumor was achieved by application of electric pulses ad libitum access to food and water. Mice (20–21 g) were used in the experiments after eight days of adaptation. Subcutaneous tumors were induced by subcutaneous injection of 2 × 106 mammary carcinoma TS/A cells in 50 µl of NaCl solution  in the left rear leg. The growth of tumors was followed by measurement of three orthogonal diameters (a × b × c) with a Vernier caliper. After six to seven days, when tumors grew to a volume of approximately 40 mm3, they were subjected to treatment.All animal experiments were conducted in accordance with the guidelines for animal experiments of the European Union directives, and permission was obtained from the Ministry of Agriculture and the Environment of the Republic of Slovenia (permission no. U34401-1/2015/7). Eleven-week-old female BALB/c mice were purchased from Charles River Laboratories Italia s.r.l.  and were maintained in specific-pathogen-free conditions with 1–m9) that were subjected to the IRE treatment, while the control group consisted of two tumors  and only had the electrodes inserted but were not IRE treated.Tumors were divided into two groups. The treated group consisted of nine tumors , xylazine  and acepromazine . IRE of mice tumors was performed by applying two sequences of four high voltage electric pulses , each of 100 µs duration with a 700 V amplitude and with a pulse repetition frequency of 5 kHz Fig. . The ele47. Next, a MREIT algorithm49  equipped with microimaging accessories with maximum gradients of 250 mT/m . The following imaging parameters were used: field of view = 30 mm; imaging matrix = 64 × 64; inter-echo delay = 2.64 ms. In the sequence  is given by:S is cell survival, E is the applied electric field, N is the number of delivered electric pulses, tp is the duration of single electric pulse, Ec is the critical electric field in which 50% of the cells are killed and k is the kinetic constant that defines the slope of the curve. Values Ec and k for different numbers and duration of electric pulses were empirically determined in vitro in a recent study on mathematical modeling of IRE-induced cell death43. For N = 8 and tp = 100 µs, which were the electric parameters of electric pulses used in our study, the values of EC and k were determined at 2.344 kV/cm and 0.2677 kV/cm, respectively43. The curve representing the probability of cell death with an increasing electric field for the determined Ec and k is shown in Fig. In this study, the Peleg-Fermi mathematical modelTo obtain maximal tissue reaction (necrosis) after IRE, a preliminary experiment to estimate the necrotic area in tumors at 48 hours, 72 hours and 96 hours post-treatment was performed . Due to the highest necrotic area observed at 72 hours after the IRE, the 72 hour timepoint was chosen for the sacrifice of mice. After the mice were sacrificed, tumors were excised and removed from underlying skin and fixed in zinc fixative . After 24 hours, the tumors were placed in 5 ml of 70% ethanol for an additional 24 h and finally embedded in paraffin. From the paraffin block in the middle of each tumor, 2 μm-thick sections were cut perpendicular to the insertion of the electrodes and stained with hematoxylin and eosin (H&E). The images of H&E-stained slides of the whole tumor were imaged with a DP72 CCD camera  connected to a BX-51 microscope . Necrotic areas in all H&E-stained slides were measured by one observer using CellSens Dimension software independently from the results obtained by MREIT.2) were calculated for different probabilities of cell death. Bland-Altman analysis52 was applied to assess the agreement between predicted IRE areas and ablated areas within limits of agreement determined as the mean differences between predicted IRE areas and ablated areas ±1.96 standard deviation . A p value of less than 0.05 was considered to indicate statistical significance. The number of animals used for experiments was defined using Pearson correlation guidelines53.All results were statistically analyzed using MATLAB 2016a and its Statistics and Machine Learning Toolbox . Linear regression was employed to assess the relationship between predicted IRE areas and ablated areas obtained by H&E staining. The slope and coefficient of determination . Two examples of electric field distribution obtained by means of MREIT are shown in Fig. The ablated zone was determined in H&E stained histological slides obtained from tumor samples 72 hours after IRE. Due to the small thickness of histological sections (2 μm), certain parts of the tumors were torn during preparation of the histological slides, especially in areas damaged by the electrodes. Images of tumor sections with nearly no torn parts are presented in Fig. AIRE) and the histologically determined ablated area (AHE) for all tumors of the treated group, linear regression was employed to calculate regression slopes and the corresponding coefficients of determination (R2) for different probabilities of cell death  and with a slope of 0.989  between the difference on the y-axis and the mean on the x-axis, suggesting that the difference is approximately constant throughout the range of values obtained at a 20% probability of cell death.Overlays of the predicted cell death probability of two tumors obtained by means of the Peleg-Fermi model with the corresponding histological slides are shown in Fig. ath Fig. . The reg989 Fig.  correspo989 Fig. . Bland-AIn this study, the MREIT technique, which enables accurate determination of the electric field distribution in a treated tumor, was complemented with a Peleg-Fermi mathematical model for predicting tissue damage due to electroporation. Thus, an efficient method for prediction of ablated tissue areas in IRE-treated tumors was obtained. The method was tested in mouse tumors, and its results showed good agreement between tumor ablated areas predicted by the tissue damage model and those confirmed by histological analysis.56. In addition to a wide variety of tissues used in the studies, this large range of reported IRE thresholds is also a consequence of different applied electric pulse parameters. The electroporation outcome is not determined solely by the electric field strength but also by the exposure time to the field, i.e., the number and duration of electric pulses41. In addition, electroporation cannot be regarded as a discrete effect where all cells exposed to an electric field above the threshold would be electroporated and those exposed to a field below the threshold would not be electroporated. In reality, there is a continuous effect with a broader transition interval of electric fields between the two states. The same characterization applies for the transition between reversibly and irreversibly electroporated tissue states. The so-called permeabilization and survival curves can thus be described using various mathematical models44. In this study, the Peleg-Fermi model was used. The model was found to be efficient in the prediction of cell survival as a function of the electric field strength51. With implementation of the survival curves to the MREIT algorithm, a more informative monitoring of IRE tissue ablation was obtained. This monitoring also enables displays of cell death probability , an established technique to reduce neurological symptoms caused by numerous movement disorders, such as Parkinson’s disease and essential tremorIn conclusion, the ability to predict irreversibly electroporated areas by means of MREIT and the Peleg-Fermi mathematical model are demonstrated. This method can potentially be used in electroporation-based clinical applications, such as IRE tissue ablation, where it can be applied for corrective interventions during the electroporation procedure and could therefore improve and assure the desired treatment outcome with complete tumor ablation. Furthermore, MREIT can provide valuable experimental results, such as IRE threshold determination and electric pulse parameters for treatment planning protocols, and for determining mathematical model parameters that can serve for prediction of damage in various tissues undergoing electroporation-based treatments.Supplementary material