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Excess mortality is often used to estimate the effect of a certain crisis on the population. It is defined as the number of deaths during a crisis exceeding the expected number based on historical trends. Here, we calculated excess mortality due to the COVID-19 pandemic for Croatia in the 2020-2021 period. The excess was calculated on the national and county level for different age and sex categories. In addition to the absolute number, the excess mortality was also expressed as a ratio of excess deaths to the predicted baseline and excess mortality rate. We showed that using both measures is necessary to avoid incorrect conclusions. The estimated excess mortality on the national level was 14,963, corresponding to an excess percentage of 14.3%. With respect to sex, there was a higher excess mortality rate for men compared to women. An exponential relationship was observed between age and the excess mortality rate.These trends wee representative of most counties as well, with large variations in the magnitude of the effect. However, there were also exceptions to the general rule. The reasons for these deviations were discussed in terms of between-county differences in demographic structure, population density and special events that took place during the pandemic.

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Despite research going back over a century showing the U.S. census counts some groups more accurately than others at the national level, little is known about how undercount varies within the country. I focus on a population easily measured with administrative data yet known to suffer high levels of undercount-native-born young children-to document state-level variation in undercount by race and Hispanic status. Although the race-specific analysis is only possible for the 2000 census, the patterns I show for all children are similar to those in 2010, implying the results from 2000 are likely relevant to today. Undercount levels vary widely across states, with non-Black children having the highest rates in the south and southwest, and Black children in the northeast. Results by Hispanic status show non-Black Hispanic young children are highly undercounted in several states with high Hispanic populations, but not all, and are also highly undercounted in the northeast and New England. In several states with high non-Black Hispanic undercounts, non-Black non-Hispanic children are also undercounted at a high rate. I find a very strong correlation between the fraction of births to foreign-born mothers in the state and the undercount of Black and non-Black children-in fact, it is the strongest correlate with the undercount of native-born Black children of those I investigate. The fraction of foreign-born mothers does not correlate with the undercount of non-Black Hispanic and non-Hispanic young children, although Hispanic status of the parents do. My results suggest a group-specific, local focus for future work is needed to determine the causes of census undercount.

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OBJECTIVE: To measure the Medicaid undercount and analyze response error in the 2007-2011 Current Population Survey Annual Social and Economic Supplement (CPS ASEC). DATA SOURCES/STUDY SETTING: Medicaid Statistical Information System (MSIS) 2006-2010 enrollment data linked to the 2007-2011 CPS ASEC person records. STUDY DESIGN: By linking individuals across datasets, we analyze false-negative error and false-positive error in reports of Medicaid enrollment. We use regression analysis to identify factors associated with response error in the 2011 CPS ASEC. PRINCIPAL FINDINGS: We find that the Medicaid undercount in the CPS ASEC ranged between 22 and 31% from 2007 to 2011. In 2011, the false-negative rate was 40%, and 27% of Medicaid reports in CPS ASEC were false positives. False-negative error is associated with the duration of enrollment in Medicaid, enrollment in Medicare and private insurance, and Medicaid enrollment in the survey year. False-positive error is associated with enrollment in Medicare and shared Medicaid coverage in the household. CONCLUSIONS: Survey estimates of Medicaid enrollment and estimates of the uninsured population are affected by both false-positive response error and false-negative response error, and these response errors are non-random.

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Privatization has grown exponentially, both in salience and in form, over the past several decades. This shifting of administrative authority away from the state can make it difficult for program recipients to link their use of a federal program back to government, a disconnect known as "submerging" the state. However, privatization is a process that occurs in degrees, and not all privatization initiatives look alike. This study leverages variation in the implementation of Medicaid managed care, which is the most widespread form of Medicaid privatization, to examine how privatization maps onto state submersion and affects state visibility. This analysis shows that, although Medicaid managed care enrollment, at large, does not relate to recipients' self-reported Medicaid enrollment, when privatized Medicaid plans introduce administrative designs that obscure the role of government, Medicaid self-reporting declines. These findings demonstrate that policy recipients are less able to recognize both the personal relevance of a specific public program and the public nature of this interaction when privatized programs utilize design features that attenuate signals of government involvement. In highlighting this disconnect, this article shows how privatization makes it more difficult for policy recipients to engage in the civic sphere as informed advocates for their self-interest.

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OBJECTIVES: To provide evidence of the magnitude of census undercounts of 'hard-to-reach' subpopulations and to improve estimation of the size of the urban indigenous population in Toronto, Canada, using respondent-driven sampling (RDS). DESIGN: Respondent-driven sampling. SETTING: The study took place in the urban indigenous community in Toronto, Canada. Three locations within the city were used to recruit study participants. PARTICIPANTS: 908 adult participants (15+) who self-identified as indigenous (First Nation, Inuit or Métis) and lived in the city of Toronto. Study participants were generally young with over 60% of indigenous adults under the age of 45 years. Household income was low with approximately two-thirds of the sample living in households which earned less than $C20 000 last year. PRIMARY AND SECONDARY OUTCOME MEASURES: We collected baseline data on demographic characteristics, including indigenous identity, age, gender, income, household type and household size. Our primary outcome asked: 'Did you complete the 2011 Census Canada questionnaire?' RESULTS: Using RDS and our large-scale survey of the urban indigenous population in Toronto, Canada, we have shown that the most recent Canadian census underestimated the size of the indigenous population in Toronto by a factor of 2 to 4. Specifically, under conservative assumptions, there are approximately 55 000 (95% CI 45 000 to 73 000) indigenous people living in Toronto, at least double the current estimate of 19 270. CONCLUSIONS: Our indigenous enumeration methods, including RDS and census completion information will have broad impacts across governmental and health policy, potentially improving healthcare access for this community. These novel applications of RDS may be relevant for the enumeration of other 'hard-to-reach' populations, such as illegal immigrants or homeless individuals in Canada and beyond.

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BACKGROUND: Little research has been done to identify reasons employers fail to report some injuries and illnesses in the Bureau of Labor Statistics Survey of Occupational Injuries and Illnesses (SOII). METHODS: We interviewed the 2012 Washington SOII respondents from establishments that had failed to report one or more eligible workers' compensation claims in the SOII about their reasons for not reporting specific claims. Qualitative content analysis methods were used to identify themes and patterns in the responses. RESULTS: Non-compliance with OSHA recordkeeping or SOII reporting instructions and data entry errors led to unreported claims. Some employers refused to include claims because they did not consider the injury to be work-related, despite workers' compensation eligibility. Participant responses brought the SOII eligibility of some claims into question. CONCLUSION: Systematic and non-systematic errors lead to SOII underreporting. Insufficient recordkeeping systems and limited knowledge of reporting requirements are barriers to accurate workplace injury records.

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CONTEXT: Unintended pregnancy remains a major public health concern in the United States. Information on pregnancy rates among contraceptive users is needed to guide medical professionals' recommendations and individuals' choices of contraceptive methods. METHODS: Data were taken from the 1995 National Survey of Family Growth (NSFG) and the 1994-1995 Abortion Patient Survey (APS). Hazards models were used to estimate method-specific contraceptive failure rates during the first six months and during the first year of contraceptive use for all U.S. women. In addition, rates were corrected to take into account the underreporting of induced abortion in the NSFG. Corrected 12-month failure rates were also estimated for subgroups of women by age, union status, poverty level, race or ethnicity, and religion. RESULTS: When contraceptive methods are ranked by effectiveness over the first 12 months of use (corrected for abortion underreporting), the implant and injectables have the lowest failure rates (2-3%), followed by the pill (8%), the diaphragm and the cervical cap (12%), the male condom (14%), periodic abstinence (21%), withdrawal (24%) and spermicides (26%). In general, failure rates are highest among cohabiting and other unmarried women, among those with an annual family income below 200% of the federal poverty level, among black and Hispanic women, among adolescents and among women in their 20s. For example, adolescent women who are not married but are cohabiting experience a failure rate of about 31% in the first year of contraceptive use, while the 12-month failure rate among married women aged 30 and older is only 7%. Black women have a contraceptive failure rate of about 19%, and this rate does not vary by family income; in contrast, overall 12-month rates are lower among Hispanic women (15%) and white women (10%), but vary by income, with poorer women having substantially greater failure rates than more affluent women. CONCLUSIONS: Levels of contraceptive failure vary widely by method, as well as by personal and background characteristics. Income's strong influence on contraceptive failure suggests that access barriers and the general disadvantage associated with poverty seriously impede effective contraceptive practice in the United States.

PIP: This study estimated method-specific contraceptive failure rates in the US. Estimates were adjusted for underreporting of induced abortion in the main survey. The correction made a sizeable impact, as 25% of the 2,157,473 conceptions due to contraceptive failure were aborted. Data were obtained from the 1995 National Survey of Family Growth and the 1994-95 Abortion Patient Survey. Analysis was based on hazard models for failure in the first 6 and 12 months. Data include 7276 contraceptive use segments. The mean duration was 9.6 months. The pill and condom had the largest shares of use segments. The lowest failure rates were for implants and injectables (2-3%). Failure rates were as follows: oral pills (8%), diaphragm and cervical cap (12%), male condom (14%), periodic abstinence (21%), withdrawal (24%), and spermicides (26%). Failure rates were highest among cohabiting and other unmarried women; women with an annual family income below 200% of the federal poverty level; among Black and Hispanic women; and among adolescents and women in their 20s. The failure rate among low income women declined during 1988-95. Women above the 200% of poverty level had stable rates. Poverty continued to have a negative impact on effective contraceptive use. Four models were used to examine the effects of socioeconomic factors on contraceptive failure.

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"As a result of lessons learnt from the 1991 [U.K.] census, a research programme was set up to seek improvements in census methodology. Underenumeration has been placed top of the agenda in this programme, and every effort is being made to achieve as high a coverage as possible in the 2001 census. In recognition, however, that 100% coverage will never be achieved, the one-number census (ONC) project was established to measure the degree of underenumeration in the 2001 census and, if possible, to adjust fully the outputs from the census for that undercount. A key component of this adjustment process is a census coverage survey (CCS). This paper presents an overview of the ONC project, focusing on the design and analysis methodology for the CCS. It also presents results that allow the reader to evaluate the robustness of this methodology."

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CONTEXT: Induced abortions are often severely underreported in national surveys, hampering the estimation and analysis of unintended pregnancies. To improve the level of abortion reporting, the 1995 National Survey of Family Growth (NSFG) incorporated new interview and self-report procedures, as well as a monetary incentive to respondents. METHODS: The weighted numbers of abortions reported in the main interview of the 1995 NSFG (Cycle 5), in the self-report and in the two procedures combined are compared with abortion estimates from The Alan Guttmacher Institute. The Cycle 5 estimates are also compared with estimates from previous cycles of the NSFG. RESULTS: The self-report produces better reporting than the main interview, but combining data from the two procedures yields the highest count of abortions. For the period 1991-1994, the level of reporting is 45% in the main interview, 52% in the self-report and 59% when the two methods are combined. The level of abortion reporting in the combined data ranges from 40% for women with an income less than the federal poverty level to more than 75% among women who were older than 35, those who were married at the time of their abortion and those with an income above 200% of the poverty level. The completeness of abortion reporting in the main interview of Cycle 5, though indicating a remarkable improvement over reporting in Cycle 4, is comparable to the levels in Cycles 2 and 3. CONCLUSIONS: The usefulness of the NSFG remains extremely limited for analyses involving unintended pregnancy and abortion.

PIP: This study assessed the extent of full reporting of induced abortion in the 1995 National Survey of Family Growth (NSFG). NSFG has new interview and self-report procedures for correcting undercounts. NSFG Cycles 2-4 were found to record under 50% of abortions that actually occurred in the US. This study compared the level of abortions reported under each of two NSFG survey procedures (the main interview and the self-report). These 2 sources were used to derive the best abortion estimates available from Cycle 5. Estimates from Cycle 5 were compared to earlier cycles for general reporting and for reporting on subgroups of women. Self-reported abortion data appeared to be somewhat less consistent than main interviews and lacked valid dates. The computer entries could increase the potential for input errors. Analysis includes the comparison between: 1) the number of abortions that were reported in the main interview with those that actually occurred in the US; 2) self-reported abortions with external estimates; and 3) both sources of abortions with external estimates. In general, women reported abortions more completely in self-reports during 1976-90. A more complete count occurred with the combined sources. Even with the new procedures in Cycle 5, abortions were undercounted. The combined sources yielded 64% of the actual abortion events. Abortion reporting in the main interview and combined sources varied widely across subgroups. However, for some subgroups, self-reports improved reporting by 33%. Higher level of education was associated with a low accuracy of reporting.

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"The 1996 Census count of indigenous Australians included a substantial number of individuals who were not recorded as indigenous by the previous census. This paper considers the implications of this for interpreting change in employment numbers. Two adjustments are made to employment change data. First, reverse survival of the 1996 population is applied to reconstruct 1991 employment figures. Second, administrative data are used to discount employment generated by participation in labour market programs. The effect is to substantially deflate the strong intercensal employment growth apparent from census counts with the conclusion that the rate of indigenous employment in the mainstream labour market has fallen."

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The author discusses the debate on how to reduce the undercount error in the 2000 U.S. census. He reviews the history of the census, and considers the impact of the "opposing ¿interest groups' represented in Congress and elsewhere debating methodological issues that can affect the distribution of census-allocated resources [who] tend to favor methodological choices that will increase their resources."

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"On the night of the 1996 Census, 5.4 per cent of the people counted in Australia were staying away from home (visitors), compared to 4.6 per cent in 1986. Queensland's share of the nation's visitors has risen since 1986, largely at the expense of Western Australia's. Queensland is now the most popular destination for interstate and overseas visitors, while New South Wales hosts a large proportion of intrastate visitors. But the overall picture is partly obscured because the Census missed around 100,000 overseas visitors."

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PIP: This article discusses the Census of India for 2001 and gender issues such as the sex ratio, undercounts of girls, tribe and scheduled caste data, and the data users' conference. The April 1998 conference was attended by representatives of government planning and development departments, research institutions, and statisticians and researchers. Under consideration was a government Ministry of Welfare proposal to reintroduce, after a 60-year hiatus, data by scheduled caste and tribe. This issue is complicated by criteria that vary by state and includability. Quotas for backward classes in educational institutions, government jobs, and in decision-making bodies are the reason for designations by caste. Some groups are distressed because of lack of inclusion as backward classes. M.N. Srinivas strongly advises that counting by caste will create problems for enumerators and will result in lawsuits and violent disturbances. G. Shah argues that caste counts will not weaken the caste system nor expand their political power, but will intensify internal conflicts between the Dalit and Bahujan movements. One other issue is the reintroduction of the 1961 Household schedule which provides family composition by landholding size and household enterprise. Krishnaji advises that this data would help analyze sex ratio imbalances. In 1997, the Core Group examined gender equity issues and operational issues about increasing the scope without increasing costs. The Core Group recommended caste data at the district and sub-district level. Undercounting of the female work force is a continuing issue from the 1991 Census. Suggestions by the Core Group are indicated in brief.^ieng

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"The aim of this paper is to analyze the pattern of delayed birth registration [in Brazil] and to establish a relationship with the total and registered births in order to estimate a mathematical function that quantifies birth underregistration."

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"Many people have been moving to Guangdong Province from other parts of China since 1978. This paper analyzes the immigration issue in the province and suggests that the provincial government should handle this matter carefully and should give consideration to both national (Chinese) and regional (Guangdong Province) interests. The government of Guangdong Province should place priority on admitting a work force with higher education or higher work skills, which will promote the economic progress of the province."

PIP: This study examined migration patterns in Guangdong Province, China during 1990-95. Data were obtained from the 1% Population Sampling Survey in 1995. Findings indicate that Guangdong received 1.9 million migrants, or 2.76% of total provincial population, from other parts of China during 1990-95. The author considers migration statistics to be undercounts. It is argued that 26.6% of the total increase in provincial population during 1985-95 was due to migration, based on census figures. The government has a migration policy. Migration is attributed to overpopulation and economic factors. The gross national product (GNP) in Guangdong increased 1.4 times since 1990, to 545 billion Yuan, or 10% of China's total GNP. The average growth rate of GNP for 5 years was 19.1%. It was 6.5% higher than rates in the national plan for 1986-90. The GNP per capita, in 1995, was 8020 Yuan. Most migrants came from economically backward regions, such as Hunan, Jiangxi, Guanxi, and Sichuan provinces (76.5%). 12.2% came from Hunan, Hubei, and Guizhou provinces. In 1994, GNP per capita in eastern China was 2.21 times as high as in central China, and 3.23 times as high as in western China. Eastern China has greater productivity. 80.4% of migrants were aged 15-39 years. Over 80% had a middle school graduation or lower educational status. 90% of migrants were blue-collar workers. 14.4% of industry and service workers were skilled workers. Guangdong is shifting to a market orientation and intensive farming. Long-term development will need investment in education in the sciences, technology, and engineering.

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PIP: This study aimed to correct errors in the estimation of the change in population size in China, during the Qing Dynasty (1741-1799). Data were obtained from household census records, the "True Facts of the Qing Dynasty" (Qing Shi Lu), and other historical records of minority, religious, and military population groups. Census records after 1740 included the number of household members instead of registered adult males aged 16-59. During 1741-75, the annual average growth rate was 13.2%. The growth rate during 1742-43 was 15.9% higher than the national average, a measure of undercounts. Underreporting was considered a problem throughout the entire study period. Undercounts in Gansu, Jiangxi, and Hubei provinces are described. Rates and underreporting varied between 30-50% in provinces. The annual average geometric growth rate during 1741-75 was an estimated 14.22%. The growth rate was 8.7% during 1776-94. It is unlikely that growth rates declined in the latter part of the century. The author adjusts figures after 1776 for undercounts of 5.52% annually. The total population undercount of 8 Manshu Banners (9.5 million), other ethnic minorities not registered (6 million), and monks, nuns, priests, and soldiers of the Green Standard (3.6 million) is an estimated 19.1 million. The adjusted total population is 388,150,057 in 1799. The annual growth rate of 14% during 1741-99 was much higher than the global growth rate of 3-5%. The 14% growth rate is likely when the birth rate is 35-38% and the mortality rate is around 20%, rates which agree with 20th century rates. Population growth in the 18th century had far reaching implications for today's population size.^ieng

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PIP: The authors evaluate the underregistration of mortality in Latin America, using the example of Ecuador. Underregistration by province, age groups, sex, and cause of death is investigated.^ieng

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The gradual increase in the sex ratio at birth (SRB) that has been recorded in China since 1980 is examined, and an attempt is made to identify the reasons for this increase. The author concludes that "such demographic variables as child-bearing age, parity and the sex of children have not determined the rising SRB since the 1980s in China. Rather, the traditional bias towards giving birth and reporting male offspring was responsible for the high SRB score and for the disparities in SRB between different socio-economic groups." The extent to which the birth of female babies is not reported is assessed, as is the extent of sex-selective abortion.

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Inaccuracy in women's reports of their abortion histories affects many areas of interest to reproductive health professionals and researchers. The identification of characteristics that affect the accuracy of reporting is essential for the improvement of data collection methods. A comparison of the medical records of 104 American women aged 27-30 in 1990-1991 with their self-reported abortion histories revealed that 19% of these women failed to report one or more abortions. Results of logistic regression analysis indicate that nonwhite women were 3.3 times as likely as whites to underreport. With each additional year that had elapsed since the first recorded abortion, women became somewhat more likely to underreport (odds ratio of 1.3), while each additional year of a woman's education slightly decreased the likelihood of underreporting (odds ratio of 0.7).

PIP: Underreporting of abortion, prevalent both between and within populations, threatens to undermine the validity of research in this area. Estimates of abortion, contraceptive failure, miscarriage, and other reproductive history events that rely on self-reports are vulnerable to bias. This study sought to identify the characteristics associated with the likelihood of underreporting an abortion history. 651 respondents from the Child Health and Development Studies who were 27-30 years old in 1990-91 filled out self-administered reproductive health questionnaires; these results were then compared to the women's 1980-92 medical records. The final sample consisted of 104 women with a medical record of a total of 170 abortions. 90% of women who had at least one abortion in their medical record reported at least one abortion; only 56% reported all of their abortions within 1 year of the medical record date, however. Overall, 19% underreported their abortion history. Of these, half reported no abortions and half reported fewer abortions than were in the medical record. Another 46% reported the exact number of abortions, and 35% reported more abortions than were in the record. Non-White women were 3.3 times more likely than Whites to underreport abortion. For each year that elapsed since the procedure, the odds of underreporting increased by 26%. Finally, every additional year of schooling decreased the likelihood of underreporting abortion by 30%.

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The aim of this study was to determine the extent to which selective under-coverage of births to mothers more likely to be at risk of HIV-1 infection will result in a significant under-estimation of the true neonatal seroprevalence. Census data, local birth statistics, maternity data and data from the prevalence monitoring programme were used to produce a model to predict the effects of under-coverage in the uptake of neonatal metabolic screening which has been observed in babies with a mother of ethnic group black African. The adjustment factor which allows for under-coverage is the relative inclusion ratio (RIR); the probability that samples from a group at different risk of HIV infection were included in the survey divided by the probability of inclusion for samples from all other babies. The RIR was found to be close to unity (0.97), indicating a minimal bias. Under usual conditions only if the relative inclusion ratio (RIR) declined to values of 0.87 or below would there be a substantial bias. Despite some selective under representation, the results obtained from the Unlinked Anonymous HIV Monitoring Programme Dried Blood Spot Survey would seem to identify levels of prevalence in the population of child-bearing women with a good degree of accuracy and remains a useful tool for resource allocation, planning of services, provision of care and counselling.

PIP: Census data, local fertility statistics, maternity data, and data from the HIV-1 prevalence monitoring program were used to develop a model to predict the effects of under-coverage of neonatal dried spot screening, which has been observed in infants with mothers of the Black African ethnic group. The data and statistics were limited to South London, England, which included the boroughs of Lambeth, West Lambeth, Southwark, and Camberwell. The model assumed HIV-1 prevalence values for African mothers to be 0-20% and 0-0.5% for all other mothers. It assumed a general fertility rate of 160/1000 for African mothers and 64/1000 for all other mothers. Coverage for neonatal dried spot screening was assumed to be 95.6% for African mothers and 98.15% for all other mothers. The relative inclusion ratio (RIR) (the probability that samples from a group at different risk of HIV infection were included in the survey divided by the probability of inclusion for samples from all other infants) was used as the adjustment factor allowing for under-coverage. The RIR was close to unity (0.975) (i.e., 95.6% actual coverage for African mothers in South London divided by 98.15% for all other mothers in South London), suggesting minimal bias. Under usual conditions, substantial bias would not occur unless RIR fell to values of 0.87 (i.e., coverage of 86% divided by 98.15%) or below. When HIV-1 prevalence was assumed to be 0.3% in all others, there was no difference between true and observed HIV-1 prevalence until it reached 3% in African mothers. At 3% HIV-1 prevalence for African mothers and 0.3% for all others, the difference was only 0.01% (i.e., observed prevalence of 0.34% compared to a true prevalence of 0.35%). These findings suggest that the Unlinked Anonymous HIV Monitoring Programme Dried Blood Spot Survey appears to identify accurately levels of HIV prevalence in the population of reproductive age women. It is still a useful tool for resource allocation, planning of services, provision of care, and counseling.

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