Female sex as a risk factor for in-hospital mortality among children undergoing cardiac surgery

Ruey-Kang R. Chang, Alex Y. Chen and Thomas S. KlitznerUndergoing Cardiac Surgery

Female Sex as a Risk Factor for In-Hospital Mortality Among Children

ISSN: 1524-4539 Copyright © 2002 American Heart Association. All rights reserved. Print ISSN: 0009-7322. Online

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Female Sex as a Risk Factor for In-Hospital MortalityAmong Children Undergoing Cardiac Surgery

Ruey-Kang R. Chang, MD, MPH; Alex Y. Chen, MD; Thomas S. Klitzner, MD, PhD

Background—The purpose of this study was to investigate whether sex disparity in cardiovascular outcomes exists inchildren who undergo cardiac surgery.

Methods and Results—Statewide hospital discharge data from California from 1995 to 1997 were used. Children �21years old who had a procedure code (by ICD9-CM) that indicated cardiac surgery were selected. The outcome variablewas binary, in-hospital death versus alive at discharge. Twenty-three surgical procedures were selected and adjusted forrisk by procedure type. We used logistic regression analysis to evaluate the effect of sex on in-hospital mortality,controlling for age, race and ethnicity, type of insurance, home income, type of admission, date and month of surgery,hospital case volume, and type of procedure. There were 6593 cases of cardiac surgery, with 345 in-hospital deaths(mortality rate 5.23%). Crude mortality rates for males (4.98%) and females (5.54%) were not significantly different.However, fewer females were neonates, and females had more low-risk procedures than males. Multivariate logisticregression showed that females had a higher odds ratio (OR) for mortality than males (OR 1.51, P�0.01). The OR formortality was 3.86 for neonates and 2.98 for infants compared with children aged �1 year. Low-volume hospitals hadhigher mortality rates than high-volume hospitals (OR 1.67, P�0.01). The risk-adjusted length of hospital stay andcharges were similar between females and males.

Conclusions—For children undergoing cardiac surgery, female sex was associated with 51% higher odds of death thanmale sex. The mechanism by which female sex acts as a risk factor requires further investigation. (Circulation. 2002;106:1514-1522.)

Key Words: sex � pediatrics � mortality � surgery

In adults undergoing CABG surgery, many studies havereported higher mortality rates in women.1–3 The risk-

adjusted odds ratio (OR) for mortality for females comparedwith males is generally reported to be 1.5 to 2.5.4–6 Wein-traub et al7 reported 13 368 patients undergoing CABG at asingle institution and found an OR for risk-adjusted mortalityof 1.79 for females compared with males. O’Connor et al8

reported an OR of 2.23 for females. Although the finding ofhigher mortality for females undergoing cardiac surgery hasbeen reported repeatedly, the mechanism by which sex acts asan independent risk factor for adult coronary artery diseaseremains unclear. Investigators have postulated that outcomedifferences by sex are due to biological factors, referral bias,and differences in utilization of services.9–11 It is unknownwhether a similar sex disparity in cardiovascular outcomesexists among children undergoing surgery for congenitalheart disease (CHD).

We hypothesized that sex is an important determinant forsurgical outcomes of children with CHD. We further hypoth-

esized that the causes for the sex difference in outcomesinclude biological factors and factors related to healthcareaccess and utilization. The objectives of the present studywere to determine whether the risk-adjusted in-hospital mor-tality rate is higher in female pediatric patients than in malepatients undergoing cardiac surgery and to explore the pos-sible causes for sex difference in cardiac outcomes inchildren.

MethodsData SourcesAbstracted hospital discharge data from the California Office ofStatewide Health Planning and Development (OSHPD) were used toconduct the analysis. The OSHPD database includes all dischargesfrom �500 acute-care hospitals in California. In the present study,we used OSHPD data on hospital discharges from January 1995 toDecember 1997. The OSHPD database contains International Clas-sification of Diseases, Ninth Revision, Clinical Modification (ICD9-CM) discharge diagnosis and procedure codes assigned by Californiahospitals to each individual discharge. Fields are provided for up to24 diagnoses and 20 procedures. Routine demographic and admin-

Received May 6, 2002; revision received June 19, 2002; accepted June 25, 2002.From the Division of Cardiology (R.-K.R.C.), Department of Pediatrics, Harbor-UCLA Medical Center, Torrance, Calif; Division of Research on

Children, Youth and Families (A.Y.C.), Department of Pediatrics, Children’s Hospital of Los Angeles, Los Angeles, Calif; and Department of Pediatrics(T.S.K.), David Geffen School of Medicine at UCLA, Los Angeles, Calif.

Presented in abstract form at the meeting of the American Academy of Pediatrics, Chicago, Ill, October 18, 2000.Correspondence to Ruey-Kang R. Chang, MD, MPH, Division of Cardiology, Department of Pediatrics, Harbor-UCLA Medical Center, 1000 W

Carson St, Torrance, CA 90509. E-mail [email protected]© 2002 American Heart Association, Inc.

Circulation is available at http://www.circulationaha.org DOI: 10.1161/01.CIR.0000029104.94858.6F

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istrative information such as age, sex, race, admission type andsource, discharge status, length of hospitalization, and total hospitalcharges are listed in the OSHPD database.

Case SelectionWe selected pediatric patients (�21 years of age) with procedurecodes in the database that indicated 1 or more cardiac surgicalprocedures listed in Table 1. Patients undergoing closure of patentductus arteriosus were excluded from the present study.

Study Variables and Risk AdjustmentIn-hospital death, identified by the discharge status in the OSHPDdatabase, was used as the main outcome variable. To comparedifferences in mortality between males and females, we used amultivariate regression analysis that included patient-related, health-care system–related, and medical variables as independent variables.Each independent variable was divided into mutually exclusivesubgroups to characterize each patient undergoing surgery. In addi-tion to sex, the patient-related variables were age, race and ethnicity,type of insurance, and family income (by median household incomeof the home zip code). Selected healthcare system–related variableswere month of surgery, day of surgery, type of admission, source ofadmission, and hospital case volume of pediatric cardiac surgery.The subgroups for each of these variables are shown in Table 2.

Medical variables included in the study included both the type ofsurgical procedure performed and the presence or absence of 1 ormore of 4 comorbid conditions. Pediatric cardiac surgery comprisesa variety of distinctly different procedures, and many procedureshave a small number of cases, insufficient to allow multivariateanalysis to be conducted on each surgical group individually.Previous investigators used different strategies to group commonpediatric cardiac procedures into 4 risk categories primarily definedby expert opinion.12,13 However, these risk-adjustment strategies

may not adequately account for the complexity and variety of cardiacsurgical procedures in children. Recently, Erickson et al14 proposeda risk-adjustment strategy that includes the majority of pediatriccardiac surgical procedures (16 procedure groups) as independentvariables in a multivariate analysis. This approach may be appealing,because the medical risk for patients in each procedure group is morehomogeneous. We modified Erickson’s risk-adjustment strategy byexpanding to 23 mutually exclusive procedure groups to further improvethe homogeneity of the cases within each group while maintaining anadequate number of cases in each group to allow for a meaningfulmultivariate analysis. These groups are listed in Table 2.

To determine the type of procedure each patient underwent, weexamined the principal procedure and other procedures listed in eachpatient’s discharge data. Although in some cases there were manyother procedures listed in the database, we examined the first 4 otherprocedures to standardize the selection process. In addition, weexamined the principal diagnosis and other diagnoses to furtherverify the type of procedure each patient underwent and to eliminatecases with ambiguous coding and coding errors. Each patient wasassigned to 1 of the 23 procedure groups. If a patient had a procedurecode for atrial septal defect (ASD) closure and a procedure code forany of the other procedures listed in Table 1, the patient was assignedto the non-ASD procedure group. Among patients with ventricularseptal defect (VSD) closure, if a procedure code was found forexcision of subaortic membrane, infundibulectomy, and coarctationrepair, the patient was listed in the complex VSD group. Patientsundergoing orthotopic heart transplantation were separated into 2groups: CHD as an underlying heart condition before transplantationand non-CHD (eg, cardiomyopathy). Patients undergoing aortopul-monary shunt who also had atrial septectomy for single-ventricletype of CHD were separated from the aortopulmonary shunt group.We also included 4 comorbid conditions as independent, nonproc-edural medical variables: Down syndrome (ICD9-CM code 758.0),

TABLE 1. Procedures and Corresponding ICD9-CM Codes

Procedure ICD9-CM Code(s)

Annuloplasty 35.33

Aortic valve replacement 35.21, 35.22

Aortopulmonary shunt (with and without atrial septostomy) 39.0

Arterial switch operation (with VSD and without VSD repair) 35.84

Atrial septal defect closure 35.51, 35.52, 35.61, 35.71

Atrial switch operation 35.91

Atrioventricular canal defect repair 35.54, 35.63, 35.73

Cavopulmonary shunt 39.21

Fontan operation 35.94

Infundibulectomy 35.34

Mitral valve replacement 35.23, 35.24

Norwood operation �3 months old with a diagnosis of hypoplastic left heartsyndrome (code 74.67) and a procedure code indicatingcardiopulmonary bypass (code 39.61)

Open valvuloplasty 35.10, 35.11, 35.12, 35.13, 35.14

Orthotopic heart transplant (for CHD and for non-CHD) 37.5

Right ventricle to pulmonary artery conduit 35.92

Resection of subaortic membrane/ring 35.35, 35.39

Tetralogy of Fallot repair 35.81

Thoracic vessel resection/replacement (eg, coarctation of aorta repair) 38.35, 38.45, 39.59

Total anomalous pulmonary venous return repair 35.82

Tricuspid or pulmonary valve replacement 35.25, 35.26, 35.27, 35.28

Truncus arteriosus repair 35.83

VSD repair (simple VSD and complex VSD repair) 35.53, 35.62, 35.72

CHD indicates congenital heart disease; VSD, ventricular septal defect.

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pulmonary hypertension (416.0), failure to thrive (783.4), andprematurity (765.0 and 765.1).

Data AnalysisTo determine whether sex is an independent risk factor for pediatriccardiac surgery mortality, logistic regression analysis was conducted.The sole dependent variable was a binary outcome variable: alive atdischarge versus in-hospital death. Within each independent vari-able, 1 subgroup was selected as the reference (OR�1) to calculateOR for mortality for the other subgroups relative to the referencesubgroups. The reference subgroups were as follows: male for “sex,”child (age �1 year) for “age,” white for “race/ethnicity,” publicinsurance for “type of insurance,” annual family income �$60 000for “family income,” weekday for “day of surgery,” October throughDecember for “month of surgery,” elective admission for “type ofadmission,” non–emergency room admission for “source of admis-sion,” high-volume hospital (annual case volume �100) for “cardiaccenter,” VSD closure for “type of surgical procedure,” and absenceof comorbidity for the 4 comorbidity variables. Data for each groupof variables were presented as ORs comparing each subgroup withits respective reference subgroup. For each OR, the 95% CI of theOR was calculated.

In the logistic regression, risk adjustment was accomplished bytreating each procedure as an independent variable. If a patient hadthat particular procedure performed, a value of “1” was listed; if not,value “0” was listed. Each patient had only 1 value of “1” listedamong the 23 procedure groups. The Hosmer and Lemeshow

TABLE 2. (Continued)

Variable n% ofTotal

Deaths,n

MortalityRate, %

Type of procedure

ASD closure 1038 15.74 2 0.19

VSD closure 1002 15.20 20 2.00

OHT for CHD 30 0.46 1 3.33

Thoracic vessel procedures 342 5.19 12 3.51

Aortopulmonary shunt 485 7.36 28 5.77

Tricuspid or pulmonary valve 210 3.19 3 1.43

OHT for non-CHD 81 1.23 3 3.70

Arterial switch operation 191 2.90 19 9.95

Complex VSD 122 1.85 4 3.28

Glenn shunt 290 4.40 12 4.14

Atrioventricular canal repair 405 6.14 21 5.19

Truncus arteriosus repair 70 1.06 10 14.29

TAPVR repair 219 3.32 26 11.87

Arterial switch operation with VSDclosure

88 1.33 12 13.64

Open valvotomy 514 7.80 23 4.47

TOF repair 610 9.25 38 6.23

Aortic valve replacement 205 3.11 8 3.90

Mitral valve replacement 86 1.30 6 6.98

RV to PA conduit 127 1.93 9 7.09

Fontan operation 235 3.56 14 5.96

Atrial switch operation 67 1.02 6 8.96

Aortopulmonary shunt with atrialseptectomy

52 0.79 17 32.69

Norwood operation 124 1.88 51 41.13

OHT indicates orthotopic heart transplantation; PA, pulmonary artery; RV,right ventricle; TAPVR, total anomalous pulmonary venous return; and TOF,tetralogy of Fallot.

TABLE 2. Sociodemographic and Clinical Characteristicsof Subjects

Variable n% ofTotal

Deaths,n

MortalityRate, %

Total 6593 100 345 5.23

Sex

Male 3597 54.56 179 4.98

Female 2996 45.44 166 5.54

Age

Neonate 1079 16.37 178 16.50

Infant 1849 28.04 103 5.57

Child 3665 55.59 64 1.75

Race

White 2498 37.89 114 4.56

Black 454 6.89 22 4.85

Hispanic 2591 39.30 135 5.21

Asian 574 8.71 25 4.36

Other races 476 7.43 49 10.00

Insurance

Private 464 7.04 29 6.25

Public 3417 51.83 187 5.47

Managed care 2566 38.92 120 4.68

Others 146 2.21 9 6.16

Median income

�$20 000 446 6.76 20 4.48

$20 000 to $40 000 1515 22.98 73 4.82

$40 000 to $60 000 3487 52.89 179 5.13

�$60 000 1145 17.37 73 6.38

Month of operation

Jan to Mar 1599 24.25 105 6.57

Apr to Jun 1779 26.98 98 5.51

Jul to Sep 1856 28.15 90 4.85

Oct to Dec 1359 20.61 52 3.83

Day of operation

Weekday 6397 97.03 322 5.03

Weekend 196 2.97 23 11.73

Source of admission

Non–emergency room 6453 97.88 327 5.07

Emergency room 140 2.12 18 12.86

Type of admission

Elective 4754 72.11 115 2.42

Nonelective 1839 27.89 230 12.51

Cardiac center

High volume 4520 68.56 226 5.00

Low volume 2073 31.44 119 5.74

Comorbidity

Down syndrome 412 6.25 12 2.91

Pulmonary hypertension 212 3.22 23 10.85

Prematurity 67 1.02 19 28.36

Failure to thrive 197 2.99 2 1.02

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TABLE 3. Patient, System, and Medical Characteristics of Female and MalePatients in the OSHPD Database

Variable

Female(n�2996)

Male(n�3597)

M/F Ratio(1.20) Pn % Total n % Total

Age

Neonate 405 13.52 674 18.74 1.66 �0.01

Infant 849 28.34 1000 27.80 1.18

Child 1742 58.14 1923 53.46 1.10

Race and ethnicity

White 1129 37.68 1369 38.06 1.21 0.19

Black 216 7.21 238 6.62 1.10

Hispanic 1198 39.99 1393 38.73 1.16

Asian/Pacific Islander 256 8.54 318 8.84 1.24

Other races 197 6.58 279 7.76 1.42

Insurance

Private 208 6.94 256 7.12 1.23 0.93

Managed care 1564 52.20 1853 51.52 1.18

Other insurance 1161 38.75 1405 39.06 1.21

Public 63 2.10 83 2.31 1.32

Median income

�$20 000 521 17.39 624 17.35 1.20 0.41

$20 000 to $40 000 1572 52.47 1912 53.16 1.22

$40 000 to $60 000 682 22.76 833 23.16 1.22

�$60 000 221 7.38 228 6.34 1.03

Day of operation

Weekend 752 25.10 847 23.55 1.13 0.52

Weekday 816 27.24 963 26.77 1.18

Month of operation

Jan to Mar 752 25.10 847 23.55 1.13 0.30

Apr to Jun 816 27.24 963 26.77 1.18

Jul to Sep 814 27.17 1042 28.97 1.28

Oct to Dec 614 20.49 745 20.71 1.21

Source of admission

Emergency room 77 2.57 119 3.31 1.55 0.09

Non–emergency room 2919 97.43 3478 96.69 1.19

Type of admission

Nonelective 743 24.80 1093 30.39 1.47 �0.01

Elective 2253 75.20 2504 69.61 1.11

Cardiac center

Low volume 951 31.74 1122 31.19 1.18 0.65

High volume 2045 68.26 2475 68.81 1.21

Comorbidity

Down syndrome 223 7.44 189 5.25 0.85 �0.01

Pulmonary hypertension 111 3.70 101 2.81 0.91 0.04

Prematurity 29 0.97 38 1.06 1.31 0.82

Failure to thrive 123 4.11 74 2.06 0.60 �0.01

Type of procedure

ASD closure 620 20.69 418 11.62 0.67 �0.01

VSD closure 463 15.45 539 14.98 1.16

OHT for CHD 10 0.33 20 0.56 2.00

Thoracic vessel procedures 135 4.51 207 5.75 1.53

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goodness-of-fit test was conducted to determine how the regressionmodel fit observed data.15 The regression was conducted with andwithout adjustment for clustering of cases in surgical centers by useof STATA 6.0 software (Stata Corp), and the results with andwithout adjustment were compared.

Additional analyses were conducted to test the hypothesis that sexdifferences in mortality result from differences in utilization ofservices between males and females. We compared the length ofhospital stay and hospital charges between males and females. Theanalysis of length of hospital stay and hospital charges was alsoadjusted for the patient, healthcare system, and medical variables bymultiple regression. A probability value �0.05 was consideredstatistically significant.

ResultsWe identified 6972 children undergoing 1 of the 23 types ofsurgical procedures for CHD in 65 hospitals in California inthe period 1995 to 1997. Although the surgeries were per-formed in 65 hospitals, 30 hospitals had only 1 pediatriccardiac surgery case over the 3-year period. After cases inhospitals with �10 cases over the study period and cases withincomplete or miscoded data were excluded, 6593 cases ofsurgery from 20 hospitals were analyzed. In this group, therewere 345 in-hospital deaths, yielding an overall mortality rateof 5.23%.

Table 2 summarizes the sociodemographic and clinicalvariables of the study subjects in the univariate (not risk-adjusted) analysis. The in-hospital mortality rate was 4.98%for males and 5.54% for females. In this analysis, females didnot appear to have a significantly higher mortality rate thanmales (OR 1.11, CI 0.90 to 1.38, P�0.33). Mortality rates

were 16.50% for neonates, 5.57% for infants, and 1.75% forchildren (�1 year). No significant difference in mortality wasfound among white, black, Hispanic, and Asian groups,although “other races” tended to have higher mortality.Patients with various types of insurance had similar mortalityrates. Among patients of different home income levels, atrend toward higher mortality rate was noted with decreasinghome income, although the differences were not statisticallysignificant.

With regard to the variable “month of operation,” we foundthe mortality rate in January through March was significantlyhigher than that in October through December (P�0.01).Surgeries performed on weekends and nonelective surgerieshad significantly higher mortality rates than those performedon weekdays or those performed after elective admissions.Similarly, patients admitted through the emergency room hadhigher mortality rates. Seven hospitals with an average annualcase volume �100 were defined as high-volume hospitals,and the remaining 13 hospitals, with annual case volume�100, were considered low-volume hospitals. The high-volume hospitals had significantly lower overall mortalityrates (5.00%) than the low-volume hospitals (5.74%).

The numbers of cases and mortality rates for the 23procedure groups used to account for medical risk are listedin Table 2. Closure of ASD was the most commonlyperformed procedure, accounting for 15.74% of all cases.This was followed by VSD closure and tetralogy of Fallotrepair. The mortality rates for the procedures ranged widely,

TABLE 3. (Continued)

Variable

Female(n�2996)

Male(n�3597)

M/F Ratio(1.20) Pn % Total n % Total

Aortopulmonary shunt 204 6.81 281 7.81 1.38

Tricuspid or pulmonary valve 92 3.07 118 3.28 1.28

OHT for non-CHD 35 1.17 46 1.28 1.31

Arterial switch 47 1.57 144 4.00 3.06

Complex VSD 61 2.04 61 1.70 1.00

Glenn shunt 121 4.04 169 4.70 1.40

Atrioventricular canal repair 219 7.31 186 5.17 0.85

Truncus arteriosus repair 35 1.17 35 0.97 1.00

TAPVR repair 86 2.87 133 3.70 1.55

Arterial switch with VSD 25 0.83 63 1.75 2.52

Open valvotomy 244 8.14 270 7.51 1.11

TOF repair 253 8.44 357 9.92 1.41

Aortic valve replacement 59 1.97 146 4.06 2.47

Mitral valve replacement 39 1.30 47 1.31 1.21

RV to PA conduit 52 1.74 75 2.09 1.44

Fontan operation 28 0.93 39 1.08 1.39

Atrial switch operation 107 3.57 128 3.56 1.20

APS with atrial septectomy 17 0.57 35 0.97 2.06

Norwood operation 44 1.47 80 2.22 1.82

APS indicates aortopulmonary shunt. All other abbreviations as in Table 2.

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fom 0.19% for ASD closure to 41.13% for the Norwoodoperation.

Univariate AnalysisThe results of comparisons of the patient, healthcare system,and medical characteristics between female and male patientsare listed in Table 3. There were 3597 males and 2996females, with a male-to-female ratio of 1.20. The age distri-butions of females and males were significantly different(P�0.01). Males represented a higher proportion of neonatesthan females. The composition of race/ethnicity, type ofinsurance, and household income level were similar betweenfemales and males. Although there were no differencesbetween males and females in day or month of admission,admission through emergency room, or surgery at low- versushigh-volume hospitals, males had a higher proportion ofnonelective surgeries than females (30.39% versus 24.80%,P�0.01). Among the 4 comorbid conditions selected, wefound that females had a higher incidence of Down syn-drome, pulmonary hypertension, and failure to thrive thanmales.

Of particular interest, there were significant differences inthe overall distribution of the procedures between males andfemales (P�0.01). On examination of the specific proceduregroups, females had a much higher proportion of ASDclosure than males (20.69% versus 11.62%), whereas maleshad more heart transplantation, aortic valve replacement, andarterial switch operation (with and without VSD closure) than

TABLE 4. Multivariate Logistic Regression Analysis of thePatient and Healthcare System–Related Variables

Variable Odds Ratio 95% CI

Sex

Female 1.51* 1.19–1.91

Male 1

Age

Neonate 3.86* 2.25–6.60

Infant 2.98* 2.03–4.38

Child 1

Race and ethnicity

Black 1.03 0.61–1.73

Hispanic 1.11 0.83–1.51

Asian 0.94 0.58–1.51

Other race 1.59* 1.07–2.37

White 1

Insurance

Private 1.64* 1.03–2.61

Managed care 1

Other insurance 1.30 0.60–2.83

Public 1.11 0.85–1.46

Median income

�$20 000 1.54 0.89–2.66

$20 000 to $40 000 1.27 0.76–2.11

$40 000 to $60 000 1.29 0.74–2.22

�$60 000 1

Day of operation

Weekend 1.18 0.71–1.96

Weekday 1

Month of operation

Jan to Mar 1.52* 1.06–2.19

Apr to Jun 1.31 0.91–1.90

Jul to Sep 1.30 0.89–1.88

Oct to Dec 1

Source of admission

Emergency room 1.59 0.87–2.91

Non–emergency room 1

Type of admission

Nonelective 2.01* 1.40–2.90

Elective 1

Cardiac center

Low volume 1.67* 1.28–2.17

High volume 1

Comorbidities

Down syndrome 0.73 0.38–1.43

Pulmonary hypertension 2.67* 1.62–4.38

Prematurity 3.00* 1.63–5.54

Failure to thrive 0.28 0.07–1.15

Type of procedure

ASD closure 0.16 0.04–0.68

VSD closure 1 � � �

OHT for CHD 1.01 0.52–1.98

TABLE 4. (Continued)

Variable Odds Ratio 95% CI

OHT for CHD 1.01 0.52–1.98

Thoracic vessel procedures 1.06 0.48–2.36

Aortopulmonary shunt 1.12 0.47–2.70

Tricuspid or pulmonary valve 1.25 0.26–5.99

OHT for non-CHD 1.43 0.38–5.45

Arterial switch operation 1.75 0.98–3.15

Complex VSD 1.80 0.52–6.22

Glenn shunt 2.41* 1.13–5.14

Atrioventricular canal repair 2.45* 1.13–5.30

Truncus arteriosus repair 2.58* 1.21–5.50

TAPVR repair 2.58* 1.04–6.43

Arterial switch operation with VSD closure 2.92* 1.07–7.94

Open valvotomy 3.01* 1.45–6.26

TOF repair 3.03* 1.81–5.08

Aortic valve replacement 3.42* 1.40–8.32

Mitral valve replacement 4.51* 1.20–17.02

RV to PA conduit 6.33* 3.54–11.30

Fontan operation 6.80* 3.13–14.77

Atrial switch operation 6.82* 2.32–19.97

Aortopulmonary shunt with atrial septectomy 7.63* 2.08–28.00

Norwood operation 11.66* 6.28–21.64

Abbreviations as in Table 2.The subgroup with an odds ratio (OR) of 1 was used as the reference group

for the variable.*Statistical significance, P�0.05.

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females. In addition, procedure groups with high moralityrates, such as aortopulmonary shunt with atrial septectomyand the Norwood operation, were more common in malesthan in females.

Logistic RegressionThe results of multivariate logistic regression analysis of allpatients, controlling for all nonsex variables, including med-ical risk as represented by the 23 procedure variables, arelisted in Table 4. The Hosmer and Lemeshow goodness-of-fittest for the logistic regression model showed that the regres-sion model fit the observed data well, with a �2 value of 4.21(P�0.84).

When all other patient, system, and medical variables werecontrolled, females had a significantly higher OR for mortal-ity than males (OR 1.51, P�0.01). Neonates had an OR formortality of 3.86 and infants had an OR of 2.98 comparedwith children �1 year old (P�0.01). Black, Asian, andHispanic patients had mortality rates similar to those forwhite patients; however, “other races” had significantlyhigher mortality (OR 1.59). A weakly positive increase in ORfor mortality was noted for private insurance patients (OR1.64) compared with children covered under managed care(P�0.04). The median income by home zip code did notappear to have a significant effect on surgical mortality,although the OR for mortality appeared to increase slightly asmedian home income decreased.

Surgeries performed on weekends carried higher mortalitythan weekday surgeries in the univariate analysis (11.73%versus 5.03%, respectively). However, with risk adjustment,the mortality OR for surgery performed on weekends was1.18, which was not significantly different from that forsurgeries performed on weekdays. Even after adjustment forrisk, the mortality rate for surgeries performed in January,February, or March was higher than that for surgeriesperformed in October, November, or December (OR 1.52,P�0.02). Surgeries performed in April, May, or June andJuly, August, or September did not appear to have greatermortality rates than those performed in October, November,or December. Admission through the emergency departmentdid not have a significant effect on mortality. Nonelectivesurgeries carried a 2-fold higher mortality than electivesurgeries (P�0.01). Finally, the OR for mortality for low-volume hospitals was significantly higher than that forhigh-volume hospitals (OR 1.67, P�0.01).

Four comorbid conditions were added to the risk adjust-ment: Down syndrome, pulmonary hypertension, failure tothrive, and prematurity. The OR for mortality increased withthe presence of pulmonary hypertension (OR 2.67, P�0.01)or if the patient was premature (OR 3.00, P�0.01). Thepresence of Down syndrome or failure to thrive did notproduce a significantly increased difference in the OR formortality.

Length of Hospital Stay and Hospital ChargesThe median length of hospital stay was 6 days for males and5 days for females (mean length of stay 11.1 days and 10.1days, respectively; P�0.01). The median hospital charge was$60 046 for males and $52 558 for females (mean hospital

charge $95 886 and $87 478, respectively; P�0.01). Therewas no statistical difference in hospital length of stay orcharges between males and females among patients who diedin the hospital.

When multiple regression was used to control for medical,patient, and healthcare system variables, we found males hadonly 0.1 day longer hospital stay than females when every-thing else was equal (P�0.77). In the multiple regressionanalysis, males had $338 less in hospital charges thanfemales, and this difference was not statistically significant(P�0.90).

DiscussionThe major finding of this study is that female sex wasassociated with higher in-hospital mortality among childrenundergoing cardiac surgery in California from 1995 to 1997.This difference is probably not due to variations in serviceutilization during hospitalization, because the risk-adjustedhospital length of stay and hospital charges were similarbetween males and females. We have also found age of thepatient and case volume of the hospital at which the surgeryis performed are important factors in determining outcomes,whereas race and ethnicity, type of insurance, and homeincome level are not.

Significant differences in mortality between males andfemales were seen after adjustment for the variations thatexist in the patients, healthcare system–related variables, andthe risks of the various types of cardiac surgery. Although thecrude mortality rate (without risk adjustment) was slightlyhigher in females than in males, with no statistically signifi-cant difference, fewer females were neonates, and femalesunderwent a larger number of low-risk procedures (such asASD closure) and a lower number of high-risk procedures(such as the Norwood operation). After adjustment for thesedifferences between females and males, the odds of dying forfemales were actually 50% higher than for males. Althoughrisk-adjustment strategies have been proposed in studies ofpediatric cardiac surgery,12–14 we have found that the ap-proach of including patient, system, and detailed procedurevariables in the risk-adjustment scheme is useful in distin-guishing patients with different levels of risk.

To the best of our knowledge, a difference in cardiacsurgical mortality between male and female has not beenreported in pediatric patients. In a study by Jenkins et al,12 sexwas not found to be a significant predictor for in-hospitaldeath by a multivariate analysis. In a study by Hannan et al,13

the risk-adjusted mortality rates for males and females werenot reported; however, the unadjusted mortality rate appearedto be slightly higher in males than in females (7.33% versus6.10%). The possible explanations for sex effect found in thepresent study but not by other authors are as follows: (1) caseselection—specifically, patent ductus arteriosus closure, alow-risk, high-prevalence surgery, which was included inother studies, was not included in the present study; (2)risk-adjustment methodologies—previous studies used 4 riskcategories to account for the variation in surgical risk com-pared with the 23 procedure groups used in the present study;and (3) different geographic areas and time frames fromwhich data were derived.

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Many researchers have proposed possible explanations forsex disparities in cardiovascular outcomes in adults. Thepostulated causes for sex differences can be broadly summa-rized as issues related to (1) biological differences betweenfemales and males, such as a higher prevalence of diabetesand hypertension,7 a lower functional class at the time ofsurgery,16 and smaller body surface area in females,8 and (2)differences in access to and use of health services, includinga lower rate of utilization for thrombolytic therapy andcardiac catheterization in females10,11 and a sex-derivedreferral bias for CABG.6,9 In the present study, we were ableto assess in-hospital healthcare utilization by analyzing risk-adjusted hospital charges, which we found to be similarbetween males and females. To the extent such information isavailable, it is unlikely that the sex difference in mortality isdue to differences in service utilization during hospitalization.Another possible explanation is that males are dischargedearlier than females, and therefore fewer male deaths occur inthe hospital. However, we found the risk-adjusted length ofhospital stay to be comparable between females and males.We also found that females have a higher incidence ofcomorbid conditions, such as Down syndrome, pulmonaryhypertension, and failure to thrive. Thus, we speculate thatthe sex difference in surgical outcomes discovered in thepresent study may be due to other biological differencesbetween young males and females that were not characterizedin this administrative data set. Further studies are clearlyneeded to provide more evidence to support our speculations.

It is not surprising that neonates and infants had highermortality rates for cardiac surgery. It is apparent from Table2 that neonates accounted for only 16.37% of all surgeries,but �50% of all in-hospital deaths occurred in neonates.Although the predominant types of cardiac surgery in neo-nates and infants differ from those in older children, riskadjustment for various types of procedures did not eliminateage-related differences in mortality. When other variableswere controlled, infants still had 2.98 times, and neonates3.86 times, higher mortality than children �1 year old. TheseORs are similar to the OR of 2.32 for �90-days-old and 1.96for 90-day-old to 1-year-old infants compared with childrenaged �1 year reported by Hannan et al.13

Recent studies have shown that children’s race, ethnicity,and insurance coverage may affect the age at which they arereferred to a cardiologist and the age at which surgery fordefinitive repair of CHD is performed.17,18 In the presentstudy, we did not find significant differences in mortalityamong whites, Hispanics, and blacks. This result is consistentwith the findings reported by Jenkins et al12 that race is not apredictor of in-hospital mortality. We had previously reportedthat Asians tended to be older at the time of complete surgicalrepair for CHD, which suggests racial or cultural differencesin access to and use of healthcare services.17 However, in thepresent study, any potential differences in access or use inAsians did not appear to lead to an increased risk for mortalitywhen undergoing cardiac surgery. In contrast, in adultsundergoing CABG surgery, race has been found to be anindependent risk factor for surgical mortality.19,20 It is inter-esting that race and ethnicity appear to affect the outcomes ofadult but not pediatric cardiac surgery. Whether the differ-

ences between adults and children with regard to the effect ofrace and ethnicity on cardiac surgery outcomes are due tobiological or sociocultural factors remains a topic for furtherinvestigation.

Bell and Redelmeier21 reported that patients with acute,serious medical conditions who are admitted to hospitals onweekends have higher mortality rates than patients admittedon weekdays. In the present study, we found the crudein-hospital mortality rate to be more than twice as high forsurgeries performed on weekends than for those performedon weekdays. However, risk adjustment eliminated this dif-ference. After risk adjustment, surgery on weekends had onlyslightly and nonsignificantly higher mortality (OR 1.18) thansurgeries performed on weekdays. The “weekend effect”reported by Bell and Redelmeier is probably less significantin the present study population because most pediatric cardiacsurgery admissions were elective (scheduled), and only asmall number of surgeries were performed on weekends(Table 2). In addition, surgeries performed on weekends weregenerally high-risk procedures, which therefore resulted in ahigher crude (unadjusted) mortality rate.

The relation between hospital case volume and favorablepatient outcomes has been demonstrated by several studiesfor a variety of surgical procedures.22–24 Specifically, acorrelation between case volume and outcomes for pediatriccardiac surgery has been reported by Jenkins et al12 andHannan et al,13 respectively. Both studies demonstrated thatrisk-adjusted, in-hospital mortality rates are lower in hospitalswith higher volumes of pediatric cardiac surgery. The studyby Jenkins et al12 found the OR for mortality was 7.7 for casesat hospitals with �10 cases/year, 2.9 for hospitals with 10 to100 cases/year, and 3.0 for hospitals with 101 to 300cases/year compared with hospitals with �300 cases/year(reference group). In the study by Hannan et al,13 the OR formortality for hospitals performing �100 cases/year was 1.42compared with hospitals with �100 cases/year. Similarly, inthe present study, we found an OR of 1.67 for low-volumehospitals (�100 cases/year) compared with high-volumehospitals.

Study LimitationsA major limitation of this study results from errors related tomiscoding and missing data in an administrative data-base.25–27 In the present study, we attempted to minimize theeffect of errors from miscoding by restricting our caseselections to hospitals with �10 cases/year of pediatriccardiac surgery. In addition, we carefully examined theprocedure and diagnosis codes for each patient. Despite thesesafeguards, missing data and miscoding of patients selectedfor the present study may exist and have the potential to biasour findings.

ConclusionsSex appears to be an important determinant of surgicaloutcome among children undergoing cardiac surgery. In thisseries of cases, females had 51% higher mortality than maleswhen other medical and nonmedical variables were heldequal. The sex difference in outcomes does not appear to berelated to in-hospital health services utilization. Future stud-

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ies to investigate the mechanisms by which sex influencessurgical outcomes may help to improve outcomes of childrenundergoing cardiac surgery.

AcknowledgmentsDr Chang was a postdoctoral fellow of the Agency for HealthcareResearch and Quality. This study was supported in part by aninstitutional grant from the Harbor-UCLA Research and EducationInstitute and grants (RR 00425 and K23 RR17041-01) from theNational Center for Research Resources, National Institutes ofHealth.

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